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EMR‐Based Detection of Deterioration
Patients who deteriorate in the hospital and are transferred to the intensive care unit (ICU) have higher mortality and greater morbidity than those directly admitted from the emergency department.[1, 2, 3] Rapid response teams (RRTs) were created to address this problem.[4, 5] Quantitative tools, such as the Modified Early Warning Score (MEWS),[6] have been used to support RRTs almost since their inception. Nonetheless, work on developing scores that can serve as triggers for RRT evaluation or intervention continues. The notion that comprehensive inpatient electronic medical records (EMRs) could support RRTs (both as a source of patient data and a platform for providing alerts) has intuitive appeal. Not surprisingly, in addition to newer versions of manual scores,[7] electronic scores are now entering clinical practice. These newer systems are being tested in research institutions,[8] hospitals with advanced capabilities,[9] and as part of proprietary systems.[10] Although a fair amount of statistical information (eg, area under the receiver operator characteristic curve of a given predictive model) on the performance of various trigger systems has been published, existing reports have not described details of how the electronic architecture is integrated with clinical practice.
Electronic alert systems generated from physiology‐based predictive models do not yet constitute mature technologies. No consensus or legal mandate regarding their role yet exists. Given this situation, studying different implementation approaches and their outcomes has value. It is instructive to consider how a given institutional solution addresses common contingenciesoperational constraints that are likely to be present, albeit in different forms, in most placesto help others understand the limitations and issues they may present. In this article we describe the structure of an EMR‐based early warning system in 2 pilot hospitals at Kaiser Permanente Northern California (KPNC). In this pilot, we embedded an updated version of a previously described early warning score[11] into the EMR. We will emphasize how its components address institutional, operational, and technological constraints. Finally, we will also describe unfinished businesschanges we would like to see in a future dissemination phase. Two important aspects of the pilot (development of a clinical response arm and addressing patient preferences with respect to supportive care) are being described elsewhere in this issue of the Journal of Hospital Medicine. Analyses of the actual impact on patient outcomes will be reported elsewhere; initial results appear favorable.[12]
INITIAL CONSTRAINTS
The ability to actually prevent inpatient deteriorations may be limited,[13] and doubts regarding the value of RRTs persist.[14, 15, 16] Consequently, work that led to the pilot occurred in stages. In the first stage (prior to 2010), our team presented data to internal audiences documenting the rates and outcomes of unplanned transfers from the ward to the ICU. Concurrently, our team developed a first generation risk adjustment methodology that was published in 2008.[17] We used this methodology to show that unplanned transfers did, in fact, have elevated mortality, and that this persisted after risk adjustment.[1, 2, 3] This phase of our work coincided with KPNC's deployment of the Epic inpatient EMR (
Once we demonstrated that we could, in fact, predict inpatient deteriorations, we still had to address medicallegal considerations, the need for a clinical response arm, and how to address patient preferences with respect to supportive or palliative care. To address these concerns and ensure that the implementation would be seamlessly integrated with routine clinical practice, our team worked for 1 year with hospitalists and other clinicians at the pilot sites prior to the go‐live date.
The primary concern from a medicallegal perspective is that once results from a predictive model (which could be an alert, severity score, comorbidity score, or other probability estimate) are displayed in the chart, relevant clinical information has been changed. Thus, failure to address such an EMR item could lead to malpractice risk for individuals and/or enterprise liability for an organization. After discussing this with senior leadership, they specified that it would be permissible to go forward so long as we could document that an educational intervention was in place to make sure that clinicians understood the system and that it was linked to specific protocols approved by hospitalists.
Current predictive models, including ours, generate a probability estimate. They do not necessarily identify the etiology of a problem or what solutions ought to be considered. Consequently, our senior leadership insisted that we be able to answer clinicians' basic question: What do we do when we get an alert? The article by Dummett et al.[20] in this issue of the Journal of Hospital Medicine describes how we addressed this constraint. Lastly, not all patients can be rescued. The article by Granich et al.[21] describes how we handled the need to respect patient choices.
PROCEDURAL COMPONENTS
The Gordon and Betty Moore Foundation, which funded the pilot, only had 1 restriction (inclusion of a hospital in the Sacramento, California area). The other site was selected based on 2 initial criteria: (1) the chosen site had to be 1 of the smaller KPNC hospitals, and (2) the chosen site had to be easily accessible for the lead author (G.J.E.). The KPNC South San Francisco hospital was selected as the alpha site and the KPNC Sacramento hospital as the beta site. One of the major drivers for these decisions was that both had robust palliative care services. The Sacramento hospital is a larger hospital with a more complex caseload.
Prior to the go‐live dates (November 19, 2013 for South San Francisco and April 16, 2014 for Sacramento), the executive committees at both hospitals reviewed preliminary data and the implementation plans for the early warning system. Following these reviews, the executive committees approved the deployment. Also during this phase, in consultation with our communications departments, we adopted the name Advance Alert Monitoring (AAM) as the outward facing name for the system. We also developed recommended scripts for clinical staff to employ when approaching a patient in whom an alert had been issued (this is because the alert is calibrated so as to predict increased risk of deterioration within the next 12 hours, which means that a patient might be surprised as to why clinicians were suddenly evaluating them). Facility approvals occurred approximately 1 month prior to the go‐live date at each hospital, permitting a shadowing phase. In this phase, selected physicians were provided with probability estimates and severity scores, but these were not displayed in the EMR front end. This shadowing phase permitted clinicians to finalize the response arms' protocols that are described in the articles by Dummett et al.[20] and Granich et al.[21] We obtained approval from the KPNC Institutional Review Board for the Protection of Human Subjects for the evaluation component that is described below.
EARLY DETECTION ALGORITHMS
The early detection algorithms we employed, which are being updated periodically, were based on our previously published work.[11, 18] Even though admitting diagnoses were found to be predictive in our original model, during actual development of the real‐time data extraction algorithms, we found that diagnoses could not be obtained reliably, so we made the decision to use a single predictive equation for all patients. The core components of the AAM score equation are the above‐mentioned LAPS2 and COPS2; these are combined with other data elements (Table 1). None of the scores are proprietary, and our equations could be replicated by any entity with a comprehensive inpatient EMR. Our early detection system is calibrated using outcomes that occurred 12 hours from when the alert is issued. For prediction, it uses data from the preceding 12 months for the COPS2 and the preceding 24 to 72 hours for physiologic data.
Category | Elements Included | Comment |
---|---|---|
Demographics | Age, sex | |
Patient location | Unit indicators (eg, 3 West); also known as bed history indicators | Only patients in general medicalsurgical ward, transitional care unit, and telemetry unit are eligible. Patients in the operating room, postanesthesia recovery room, labor and delivery service, and pediatrics are ineligible. |
Health services | Admission venue | Emergency department admission or not. |
Elapsed length of stay in hospital up to the point when data are scanned | Interhospital transport is common in our integrated delivery system; this data element requires linking both unit stays as well as stays involving different hospitals. | |
Status | Care directive orders | Patients with a comfort careonly order are not eligible; all other patients (full code, partial code, and do not resuscitate) are. |
Admission status | Inpatients and patients admitted for observation status are eligible. | |
Physiologic | Vital signs, laboratory tests, neurological status checks | See online Appendices and references [6] and [15] for details on how we extract, format, and transform these variables. |
Composite indices | Generic severity of illness score | See text and description in reference [15] for details on the Laboratory‐based Acute Physiology score, version 2 and the Comorbidity Point Score, version 2. |
Longitudinal comorbidity score |
During the course of developing the real‐time extraction algorithms, we encountered a number of delays in real‐time data acquisition. These fall into 2 categories: charting delay and server delay. Charting delay is due to nonautomated charting of vital signs by nurses (eg, a nurse obtains vital signs on a patient, writes them down on paper, and then enters them later). In general, this delay was in the 15‐ to 30‐minute range, but occasionally was as high as 2 hours. Server delay, which was variable and ranged from a few minutes to (occasionally) 1 to 2 hours, is due to 2 factors. The first is that certain point of care tests were not always uploaded into the EMR immediately. This is because the testing units, which can display results to clinicians within minutes, must be physically connected to a computer for uploading results. The second is the processing time required for the system to cycle through hundreds of patient records in the context of a very large EMR system (the KPNC Epic build runs in 6 separate geographic instances, and our system runs in 2 of these). Figure 1 shows that each probability estimate thus has what we called an uncertainty period of 2 hours (the +2 hours addresses the fact that we needed to give clinicians a minimum time to respond to an alert). Given limited resources and the need to balance accuracy of the alerts, adequate lead time, the presence of an uncertainty period, and alert fatigue, we elected to issue alerts every 6 hours (with the exact timing based on facility preferences).
A summary of the components of our equation is provided in the Supporting Information, Appendices, in the online version of this article. The statistical performance characteristics of our final equation, which are based on approximately 262 million individual data points from 650,684 hospitalizations in which patients experienced 20,471 deteriorations, is being reported elsewhere. Between November 19, 2013 and November 30, 2015 (the most recent data currently available to us for analysis), a total of 26,386 patients admitted to the ward or transitional care unit at the 2 pilot sites were scored by the AAM system, and these patients generated 3,881 alerts involving a total of 1,413 patients, which meant an average of 2 alerts per day at South San Francisco and 4 alerts per day in Sacramento. Resource limitations have precluded us from conducting formal surveys to assess clinician acceptance. However, repeated meetings with both hospitalists as well as RRT nurses indicated that favorable departmental consensus exists.
INSTANTIATION OF ALGORITHMS IN THE EMR
Given the complexity of the calculations involving many variables (Table 1), we elected to employ Web services to extract data for processing using a Java application outside the EMR, which then pushed results into the EMR front end (Figure 2). Additional details on this decision are provided in the Supporting Information, Appendices, in the online version of this article. Our team had to expend considerable resources and time to map all necessary data elements in the real time environment, whose identifying characteristics are not the same as those employed by the KPHC data warehouse. Considerable debugging was required during the first 7 months of the pilot. Troubleshooting for the application was often required on very short notice (eg, when the system unexpectedly stopped issuing alerts during a weekend, or when 1 class of patients suddenly stopped receiving scores). It is likely that future efforts to embed algorithms in EMRs will experience similar difficulties, and it is wise to budget so as maximize available analytic and application programmer resources.
Figure 3 shows the final appearance of the graphical user interface at KPHC, which provides clinicians with 3 numbers: ADV ALERT SCORE (AAM score) is the probability of experiencing unplanned transfer within the next 12 hours, COPS is the COPS2, and LAPS is the LAPS2 assigned at the time a patient is placed in a hospital room. The current protocol in place is that the clinical response arm is triggered when the AAM score is 8.
LIMITATIONS
One of the limitations of working with a commercial EMR in a large system, such as KPNC, is that of scalability. Understandably, the organization is reluctant to make changes in the EMR that will not ultimately be deployed across all hospitals in the system. Thus, any significant modification of the EMR or its associated workflows must, from the outset, be structured for subsequent spread to the remaining hospitals (19 in our case). Because we had not deployed a system like this before, we did not know what to expect and, had we known then what experience has taught us, our initial requests would have been different. Table 2 summarizes the major changes we would have made to our implementation strategy had we known then what we know now.
Component | Status in Pilot Application | Desirable Changes |
---|---|---|
| ||
Degree of disaster recovery support | System outages are handled on an ad hoc basis. | Same level of support as is seen in regular clinical systems (24/7 technical support). |
Laboratory data feed | Web service. | It would be extremely valuable to have a definite answer about whether alternative data feeds would be faster and more reliable. |
LAPS2 score | Score appears only on ward or TCU patients. | Display for all hospitalized adults (include anyone 18 years and include ICU patients). |
Score appears only on inpatient physician dashboard. | Display scores in multiple dashboards (eg, emergency department dashboard). | |
COPS2 score | Score appears only on ward or TCU patients. | Display for all hospitalized adults (include anyone 18 years and include ICU patients). |
Score appears only on inpatient physician dashboard. | Display scores in multiple dashboards (eg, emergency department dashboard). | |
Alert response tracking | None is available. | Functionality that permits tracking what the status is of patients in whom an alert was issued (who responded, where it is charted, etc.)could be structured as a workbench report in KP HealthConnectvery important because of medical legal reasons. |
Trending capability for scores | None is available. | Trending display available in same location where vital signs and laboratory test results are displayed. |
Messaging capability | Not currently available. | Transmission of scores to rapid response team (or other designated first responder) via a smartphone, thus obviating the need for staff to check the inpatient dashboard manually every 6 hours. |
EVALUATION STRATEGY
Due to institutional constraints, it is not possible for us to conduct a gold standard pilot using patient‐level randomization, as described by Kollef et al.[8] Consequently, in addition to using the pilot to surface specific implementation issues, we had to develop a parallel scoring system for capturing key data points (scores, outcomes) not just at the 2 pilot sites, but also at the remaining 19 KPNC hospitals. This required that we develop electronic tools that would permit us to capture these data elements continuously, both prospectively as well as retrospectively. Thus, to give an example, we developed a macro that we call LAPS2 any time that permits us to assign a retrospective severity score given any T0. Our ultimate goal is to evaluate the system's deployment using a stepped wedge design[22] in which geographically contiguous clusters of 2 to 4 hospitals go live periodically. The silver standard (a cluster trial involving randomization at the individual hospital level[23]) is not feasible because KPNC hospitals span a very broad geographic area, and it is more resource intensive in a shorter time span. In this context, the most important output from a pilot such as this is to generate an estimate of likely impact; this estimate then becomes a critical component for power calculations for the stepped wedge.
Our ongoing evaluation has all the limitations inherent in the analysis of nonrandomized interventions. Because it only involves 2 hospitals, it is difficult to assess variation due to facility‐specific factors. Finally, because our priority was to avoid alert fatigue, the total number of patients who experience an alert is small, limiting available sample size. Given these constraints, we will employ a counterfactual method, multivariate matching,[24, 25, 26] so as to come as close as possible to simulating a randomized trial. To control for hospital‐specific factors, matching will be combined with difference‐in‐differences[27, 28] methodology. Our basic approach takes advantage of the fact that, although our alert system is currently running in 2 hospitals, it is possible for us to assign a retrospective alert to patients at all KPNC hospitals. Using multivariate matching techniques, we will then create a cohort in which each patient who received an alert is matched to 2 patients who are given a retrospective virtual alert during the same time period in control facilities. The pre‐ and postimplementation outcomes of pilot and matched controls are compared. The matching algorithms specify exact matches on membership status, whether or not the patient had been admitted to the ICU prior to the first alert, and whether or not the patient was full code at the time of an alert. Once potential matches are found using the above procedures, our algorithms seek the closest match for the following variables: age, alert probability, COPS2, and admission LAPS2. Membership status is important, because many individuals who are not covered by the Kaiser Foundation Health Plan, Inc., are hospitalized at KPNC hospitals. Because these nonmembers' postdischarge outcomes cannot be tracked, it is important to control for this variable in our analyses.
Our electronic evaluation strategy also can be used to quantify pilot effects on length of stay (total, after an alert, and ICU), rehospitalization, use of hospice, mortality, and cost. However, it is not adequate for the evaluation of whether or not patient preferences are respected. Consequently, we have also developed manual review instruments for structured electronic chart review (the coding form and manual are provided in the online Appendix of the article in this issue of Journal of Hospital Medicine by Granich et al.[21]). This review will focus on issues such as whether or not patients' surrogates were identified, whether goals of care were discussed, and so forth. In those cases where patients died in the hospital, we will also review whether death occurred after resuscitation, whether family members were present, and so forth.
As noted above and in Figure 1, charting delays can result in uncertainty periods. We have found that these delays can also result in discrepancies in which data extracted from the real time system do not match those extracted from the data warehouse. These discrepancies can complicate creation of analysis datasets, which in turn can lead to delays in completing analyses. Such delays can cause significant problems with stakeholders. In retrospect, we should have devoted more resources to ongoing electronic audits and to the development of algorithms that formally address charting delays.
LESSONS LEARNED AND THOUGHTS ON FUTURE DISSEMINATION
We believe that embedding predictive models in the EMR will become an essential component of clinical care. Despite resource limitations and having to work in a frontier area, we did 3 things well. We were able to embed a complex set of equations and display their outputs in a commercial EMR outside the research setting. In a setting where hospitalists could have requested discontinuation of the system, we achieved consensus that it should remain the standard of care. Lastly, as a result of this work, KPNC will be deploying this early warning system in all its hospitals, so our overall implementation and communication strategy has been sound.
Nonetheless, our road to implementation has been a bumpy one, and we have learned a number of valuable lessons that are being incorporated into our future work. They merit sharing with the broader medical community. Using the title of a song by Ricky SkaggsIf I Had It All Again to Dowe can summarize what we learned with 3 phrases: engage leadership early, provide simpler explanations, and embed the evaluation in the solution.
Although our research on risk adjustment and the epidemiology was known to many KPNC leaders and clinicians, our initial engagement focus was on connecting with hospital physicians and operational leaders who worked in quality improvement. In retrospect, the research team should have engaged with 2 different communities much soonerthe information technology community and that component of leadership that focused on the EMR and information technology issues. Although these 2 broad communities interact with operations all the time, they do not necessarily have regular contact with research developments that might affect both EMR as well as quality improvement operations simultaneously. Not seeking this early engagement probably slowed our work by 9 to 15 months, because of repeated delays resulting from our assumption that the information technology teams understood things that were clear to us but not to them. One major result of this at KPNC is that we now have a regular quarterly meeting between researchers and the EMR leadership. The goal of this regular meeting is to make sure that operational leaders and researchers contemplating projects with an informatics component communicate early, long before any consideration of implementation occurs.
Whereas the notion of providing early warning seems intuitive and simple, translating this into a set of equations is challenging. However, we have found that developing equations is much easier than developing communication strategies suitable for people who are not interested in statistics, a group that probably constitutes the majority of clinicians. One major result of this learning now guiding our work is that our team devotes more time to considering existing and possible workflows. This process includes spending more time engaging with clinicians around how they use information. We are also experimenting with different ways of illustrating statistical concepts (eg, probabilities, likelihood ratios).
As is discussed in the article by Dummett et al.,[20] 1 workflow component that remains unresolved is that of documentation. It is not clear what the documentation standard should be for a deterioration probability. Solving this particular conundrum is not something that can be done by electronic or statistical means. However, also with the benefit of hindsight, we now know that we should have put more energy into automated electronic tools that provide support for documentation after an alert. In addition to being requested by clinicians, having tools that automatically generate tracers as part of both the alerting and documentation process would also make evaluation easier. For example, it would permit a better delineation of the causal path between the intervention (providing a deterioration probability) and patient outcomes. In future projects, incorporation of such tools will get much more prominence.
Acknowledgements
The authors thank Dr. Michelle Caughey, Dr. Philip Madvig, Dr. Patricia Conolly, and Ms. Barbara Crawford for their administrative support, Dr. Tracy Lieu for reviewing the manuscript, and Ms. Rachel Lesser for formatting the manuscript.
Disclosures: This work was supported by a grant from the Gordon and Betty Moore Foundation (Early Detection, Prevention, and Mitigation of Impending Physiologic Deterioration in Hospitalized Patients Outside Intensive Care: Phase 3, pilot), The Permanente Medical Group, Inc., and Kaiser Foundation Hospitals, Inc. As part of our agreement with the Gordon and Betty Moore Foundation, we made a commitment to disseminate our findings in articles such as this one. However, the Foundation and its staff played no role in how we actually structured our articles, nor did they review or preapprove any of the manuscripts submitted as part of the dissemination component. Dr. Liu was supported by the National Institute for General Medical Sciences award K23GM112018. None of the sponsors had any involvement in our decision to submit this manuscript or in the determination of its contents. None of the authors has any conflicts of interest to declare of relevance to this work
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6(2):74–80. , , , , , .
- Adverse outcomes associated with delayed intensive care unit transfers in an integrated healthcare system. J Hosp Med. 2012;7(3):224–230. , , , .
- Risk factors for unplanned transfer to intensive care within 24 hours of admission from the emergency department in an integrated healthcare system. J Hosp Med. 2012;8(1):13–19. , , , , , .
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- National Health Service. National Early Warning Score (NEWS). Standardising the Assessment Of Acute‐Illness Severity in the NHS. Report of a Working Party. London, United Kingdom: Royal College of Physicians; 2012.
- A randomized trial of real‐time automated clinical deterioration alerts sent to a rapid response team. J Hosp Med. 2014;9(7):424–429. , , , et al.
- Automated detection of physiologic deterioration in hospitalized patients. J Am Med Inform Assoc. 2015;22(2):350–360. , , , et al.
- Identifying patients at increased risk for unplanned readmission. Med Care. 2013;51(9):761–766. , , , , .
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- Early detection of impending deterioration outside the ICU: a difference‐in‐differences (DiD) study. Presented at: American Thoracic Society International Conference, San Francisco, California; May 13–18, 2016; A7614. , , , et al.
- Unplanned transfers to a medical intensive care unit: causes and relationship to preventable errors in care. J Hosp Med. 2011;6(2):68–72. , , , .
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- Rethinking rapid response teams. JAMA. 2010;304(12):1375–1376. , .
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- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated healthcare delivery system. Med Care. 2013;51(5):446–453. , , , , .
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- J Hosp Med. 2016;11:000–000. et al.
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- Comparison of multivariate matching methods: structures, distances, and algorithms. J Comput Graph Stat. 1993;2(4):405–420. , .
- A method/macro based on propensity score and Mahalanobis distance to reduce bias in treatment comparison in observational study. Eli Lilly working paper available at: http://www.lexjansen.com/pharmasug/2006/publichealthresearch/pr05.pdf. , , .
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Patients who deteriorate in the hospital and are transferred to the intensive care unit (ICU) have higher mortality and greater morbidity than those directly admitted from the emergency department.[1, 2, 3] Rapid response teams (RRTs) were created to address this problem.[4, 5] Quantitative tools, such as the Modified Early Warning Score (MEWS),[6] have been used to support RRTs almost since their inception. Nonetheless, work on developing scores that can serve as triggers for RRT evaluation or intervention continues. The notion that comprehensive inpatient electronic medical records (EMRs) could support RRTs (both as a source of patient data and a platform for providing alerts) has intuitive appeal. Not surprisingly, in addition to newer versions of manual scores,[7] electronic scores are now entering clinical practice. These newer systems are being tested in research institutions,[8] hospitals with advanced capabilities,[9] and as part of proprietary systems.[10] Although a fair amount of statistical information (eg, area under the receiver operator characteristic curve of a given predictive model) on the performance of various trigger systems has been published, existing reports have not described details of how the electronic architecture is integrated with clinical practice.
Electronic alert systems generated from physiology‐based predictive models do not yet constitute mature technologies. No consensus or legal mandate regarding their role yet exists. Given this situation, studying different implementation approaches and their outcomes has value. It is instructive to consider how a given institutional solution addresses common contingenciesoperational constraints that are likely to be present, albeit in different forms, in most placesto help others understand the limitations and issues they may present. In this article we describe the structure of an EMR‐based early warning system in 2 pilot hospitals at Kaiser Permanente Northern California (KPNC). In this pilot, we embedded an updated version of a previously described early warning score[11] into the EMR. We will emphasize how its components address institutional, operational, and technological constraints. Finally, we will also describe unfinished businesschanges we would like to see in a future dissemination phase. Two important aspects of the pilot (development of a clinical response arm and addressing patient preferences with respect to supportive care) are being described elsewhere in this issue of the Journal of Hospital Medicine. Analyses of the actual impact on patient outcomes will be reported elsewhere; initial results appear favorable.[12]
INITIAL CONSTRAINTS
The ability to actually prevent inpatient deteriorations may be limited,[13] and doubts regarding the value of RRTs persist.[14, 15, 16] Consequently, work that led to the pilot occurred in stages. In the first stage (prior to 2010), our team presented data to internal audiences documenting the rates and outcomes of unplanned transfers from the ward to the ICU. Concurrently, our team developed a first generation risk adjustment methodology that was published in 2008.[17] We used this methodology to show that unplanned transfers did, in fact, have elevated mortality, and that this persisted after risk adjustment.[1, 2, 3] This phase of our work coincided with KPNC's deployment of the Epic inpatient EMR (
Once we demonstrated that we could, in fact, predict inpatient deteriorations, we still had to address medicallegal considerations, the need for a clinical response arm, and how to address patient preferences with respect to supportive or palliative care. To address these concerns and ensure that the implementation would be seamlessly integrated with routine clinical practice, our team worked for 1 year with hospitalists and other clinicians at the pilot sites prior to the go‐live date.
The primary concern from a medicallegal perspective is that once results from a predictive model (which could be an alert, severity score, comorbidity score, or other probability estimate) are displayed in the chart, relevant clinical information has been changed. Thus, failure to address such an EMR item could lead to malpractice risk for individuals and/or enterprise liability for an organization. After discussing this with senior leadership, they specified that it would be permissible to go forward so long as we could document that an educational intervention was in place to make sure that clinicians understood the system and that it was linked to specific protocols approved by hospitalists.
Current predictive models, including ours, generate a probability estimate. They do not necessarily identify the etiology of a problem or what solutions ought to be considered. Consequently, our senior leadership insisted that we be able to answer clinicians' basic question: What do we do when we get an alert? The article by Dummett et al.[20] in this issue of the Journal of Hospital Medicine describes how we addressed this constraint. Lastly, not all patients can be rescued. The article by Granich et al.[21] describes how we handled the need to respect patient choices.
PROCEDURAL COMPONENTS
The Gordon and Betty Moore Foundation, which funded the pilot, only had 1 restriction (inclusion of a hospital in the Sacramento, California area). The other site was selected based on 2 initial criteria: (1) the chosen site had to be 1 of the smaller KPNC hospitals, and (2) the chosen site had to be easily accessible for the lead author (G.J.E.). The KPNC South San Francisco hospital was selected as the alpha site and the KPNC Sacramento hospital as the beta site. One of the major drivers for these decisions was that both had robust palliative care services. The Sacramento hospital is a larger hospital with a more complex caseload.
Prior to the go‐live dates (November 19, 2013 for South San Francisco and April 16, 2014 for Sacramento), the executive committees at both hospitals reviewed preliminary data and the implementation plans for the early warning system. Following these reviews, the executive committees approved the deployment. Also during this phase, in consultation with our communications departments, we adopted the name Advance Alert Monitoring (AAM) as the outward facing name for the system. We also developed recommended scripts for clinical staff to employ when approaching a patient in whom an alert had been issued (this is because the alert is calibrated so as to predict increased risk of deterioration within the next 12 hours, which means that a patient might be surprised as to why clinicians were suddenly evaluating them). Facility approvals occurred approximately 1 month prior to the go‐live date at each hospital, permitting a shadowing phase. In this phase, selected physicians were provided with probability estimates and severity scores, but these were not displayed in the EMR front end. This shadowing phase permitted clinicians to finalize the response arms' protocols that are described in the articles by Dummett et al.[20] and Granich et al.[21] We obtained approval from the KPNC Institutional Review Board for the Protection of Human Subjects for the evaluation component that is described below.
EARLY DETECTION ALGORITHMS
The early detection algorithms we employed, which are being updated periodically, were based on our previously published work.[11, 18] Even though admitting diagnoses were found to be predictive in our original model, during actual development of the real‐time data extraction algorithms, we found that diagnoses could not be obtained reliably, so we made the decision to use a single predictive equation for all patients. The core components of the AAM score equation are the above‐mentioned LAPS2 and COPS2; these are combined with other data elements (Table 1). None of the scores are proprietary, and our equations could be replicated by any entity with a comprehensive inpatient EMR. Our early detection system is calibrated using outcomes that occurred 12 hours from when the alert is issued. For prediction, it uses data from the preceding 12 months for the COPS2 and the preceding 24 to 72 hours for physiologic data.
Category | Elements Included | Comment |
---|---|---|
Demographics | Age, sex | |
Patient location | Unit indicators (eg, 3 West); also known as bed history indicators | Only patients in general medicalsurgical ward, transitional care unit, and telemetry unit are eligible. Patients in the operating room, postanesthesia recovery room, labor and delivery service, and pediatrics are ineligible. |
Health services | Admission venue | Emergency department admission or not. |
Elapsed length of stay in hospital up to the point when data are scanned | Interhospital transport is common in our integrated delivery system; this data element requires linking both unit stays as well as stays involving different hospitals. | |
Status | Care directive orders | Patients with a comfort careonly order are not eligible; all other patients (full code, partial code, and do not resuscitate) are. |
Admission status | Inpatients and patients admitted for observation status are eligible. | |
Physiologic | Vital signs, laboratory tests, neurological status checks | See online Appendices and references [6] and [15] for details on how we extract, format, and transform these variables. |
Composite indices | Generic severity of illness score | See text and description in reference [15] for details on the Laboratory‐based Acute Physiology score, version 2 and the Comorbidity Point Score, version 2. |
Longitudinal comorbidity score |
During the course of developing the real‐time extraction algorithms, we encountered a number of delays in real‐time data acquisition. These fall into 2 categories: charting delay and server delay. Charting delay is due to nonautomated charting of vital signs by nurses (eg, a nurse obtains vital signs on a patient, writes them down on paper, and then enters them later). In general, this delay was in the 15‐ to 30‐minute range, but occasionally was as high as 2 hours. Server delay, which was variable and ranged from a few minutes to (occasionally) 1 to 2 hours, is due to 2 factors. The first is that certain point of care tests were not always uploaded into the EMR immediately. This is because the testing units, which can display results to clinicians within minutes, must be physically connected to a computer for uploading results. The second is the processing time required for the system to cycle through hundreds of patient records in the context of a very large EMR system (the KPNC Epic build runs in 6 separate geographic instances, and our system runs in 2 of these). Figure 1 shows that each probability estimate thus has what we called an uncertainty period of 2 hours (the +2 hours addresses the fact that we needed to give clinicians a minimum time to respond to an alert). Given limited resources and the need to balance accuracy of the alerts, adequate lead time, the presence of an uncertainty period, and alert fatigue, we elected to issue alerts every 6 hours (with the exact timing based on facility preferences).
A summary of the components of our equation is provided in the Supporting Information, Appendices, in the online version of this article. The statistical performance characteristics of our final equation, which are based on approximately 262 million individual data points from 650,684 hospitalizations in which patients experienced 20,471 deteriorations, is being reported elsewhere. Between November 19, 2013 and November 30, 2015 (the most recent data currently available to us for analysis), a total of 26,386 patients admitted to the ward or transitional care unit at the 2 pilot sites were scored by the AAM system, and these patients generated 3,881 alerts involving a total of 1,413 patients, which meant an average of 2 alerts per day at South San Francisco and 4 alerts per day in Sacramento. Resource limitations have precluded us from conducting formal surveys to assess clinician acceptance. However, repeated meetings with both hospitalists as well as RRT nurses indicated that favorable departmental consensus exists.
INSTANTIATION OF ALGORITHMS IN THE EMR
Given the complexity of the calculations involving many variables (Table 1), we elected to employ Web services to extract data for processing using a Java application outside the EMR, which then pushed results into the EMR front end (Figure 2). Additional details on this decision are provided in the Supporting Information, Appendices, in the online version of this article. Our team had to expend considerable resources and time to map all necessary data elements in the real time environment, whose identifying characteristics are not the same as those employed by the KPHC data warehouse. Considerable debugging was required during the first 7 months of the pilot. Troubleshooting for the application was often required on very short notice (eg, when the system unexpectedly stopped issuing alerts during a weekend, or when 1 class of patients suddenly stopped receiving scores). It is likely that future efforts to embed algorithms in EMRs will experience similar difficulties, and it is wise to budget so as maximize available analytic and application programmer resources.
Figure 3 shows the final appearance of the graphical user interface at KPHC, which provides clinicians with 3 numbers: ADV ALERT SCORE (AAM score) is the probability of experiencing unplanned transfer within the next 12 hours, COPS is the COPS2, and LAPS is the LAPS2 assigned at the time a patient is placed in a hospital room. The current protocol in place is that the clinical response arm is triggered when the AAM score is 8.
LIMITATIONS
One of the limitations of working with a commercial EMR in a large system, such as KPNC, is that of scalability. Understandably, the organization is reluctant to make changes in the EMR that will not ultimately be deployed across all hospitals in the system. Thus, any significant modification of the EMR or its associated workflows must, from the outset, be structured for subsequent spread to the remaining hospitals (19 in our case). Because we had not deployed a system like this before, we did not know what to expect and, had we known then what experience has taught us, our initial requests would have been different. Table 2 summarizes the major changes we would have made to our implementation strategy had we known then what we know now.
Component | Status in Pilot Application | Desirable Changes |
---|---|---|
| ||
Degree of disaster recovery support | System outages are handled on an ad hoc basis. | Same level of support as is seen in regular clinical systems (24/7 technical support). |
Laboratory data feed | Web service. | It would be extremely valuable to have a definite answer about whether alternative data feeds would be faster and more reliable. |
LAPS2 score | Score appears only on ward or TCU patients. | Display for all hospitalized adults (include anyone 18 years and include ICU patients). |
Score appears only on inpatient physician dashboard. | Display scores in multiple dashboards (eg, emergency department dashboard). | |
COPS2 score | Score appears only on ward or TCU patients. | Display for all hospitalized adults (include anyone 18 years and include ICU patients). |
Score appears only on inpatient physician dashboard. | Display scores in multiple dashboards (eg, emergency department dashboard). | |
Alert response tracking | None is available. | Functionality that permits tracking what the status is of patients in whom an alert was issued (who responded, where it is charted, etc.)could be structured as a workbench report in KP HealthConnectvery important because of medical legal reasons. |
Trending capability for scores | None is available. | Trending display available in same location where vital signs and laboratory test results are displayed. |
Messaging capability | Not currently available. | Transmission of scores to rapid response team (or other designated first responder) via a smartphone, thus obviating the need for staff to check the inpatient dashboard manually every 6 hours. |
EVALUATION STRATEGY
Due to institutional constraints, it is not possible for us to conduct a gold standard pilot using patient‐level randomization, as described by Kollef et al.[8] Consequently, in addition to using the pilot to surface specific implementation issues, we had to develop a parallel scoring system for capturing key data points (scores, outcomes) not just at the 2 pilot sites, but also at the remaining 19 KPNC hospitals. This required that we develop electronic tools that would permit us to capture these data elements continuously, both prospectively as well as retrospectively. Thus, to give an example, we developed a macro that we call LAPS2 any time that permits us to assign a retrospective severity score given any T0. Our ultimate goal is to evaluate the system's deployment using a stepped wedge design[22] in which geographically contiguous clusters of 2 to 4 hospitals go live periodically. The silver standard (a cluster trial involving randomization at the individual hospital level[23]) is not feasible because KPNC hospitals span a very broad geographic area, and it is more resource intensive in a shorter time span. In this context, the most important output from a pilot such as this is to generate an estimate of likely impact; this estimate then becomes a critical component for power calculations for the stepped wedge.
Our ongoing evaluation has all the limitations inherent in the analysis of nonrandomized interventions. Because it only involves 2 hospitals, it is difficult to assess variation due to facility‐specific factors. Finally, because our priority was to avoid alert fatigue, the total number of patients who experience an alert is small, limiting available sample size. Given these constraints, we will employ a counterfactual method, multivariate matching,[24, 25, 26] so as to come as close as possible to simulating a randomized trial. To control for hospital‐specific factors, matching will be combined with difference‐in‐differences[27, 28] methodology. Our basic approach takes advantage of the fact that, although our alert system is currently running in 2 hospitals, it is possible for us to assign a retrospective alert to patients at all KPNC hospitals. Using multivariate matching techniques, we will then create a cohort in which each patient who received an alert is matched to 2 patients who are given a retrospective virtual alert during the same time period in control facilities. The pre‐ and postimplementation outcomes of pilot and matched controls are compared. The matching algorithms specify exact matches on membership status, whether or not the patient had been admitted to the ICU prior to the first alert, and whether or not the patient was full code at the time of an alert. Once potential matches are found using the above procedures, our algorithms seek the closest match for the following variables: age, alert probability, COPS2, and admission LAPS2. Membership status is important, because many individuals who are not covered by the Kaiser Foundation Health Plan, Inc., are hospitalized at KPNC hospitals. Because these nonmembers' postdischarge outcomes cannot be tracked, it is important to control for this variable in our analyses.
Our electronic evaluation strategy also can be used to quantify pilot effects on length of stay (total, after an alert, and ICU), rehospitalization, use of hospice, mortality, and cost. However, it is not adequate for the evaluation of whether or not patient preferences are respected. Consequently, we have also developed manual review instruments for structured electronic chart review (the coding form and manual are provided in the online Appendix of the article in this issue of Journal of Hospital Medicine by Granich et al.[21]). This review will focus on issues such as whether or not patients' surrogates were identified, whether goals of care were discussed, and so forth. In those cases where patients died in the hospital, we will also review whether death occurred after resuscitation, whether family members were present, and so forth.
As noted above and in Figure 1, charting delays can result in uncertainty periods. We have found that these delays can also result in discrepancies in which data extracted from the real time system do not match those extracted from the data warehouse. These discrepancies can complicate creation of analysis datasets, which in turn can lead to delays in completing analyses. Such delays can cause significant problems with stakeholders. In retrospect, we should have devoted more resources to ongoing electronic audits and to the development of algorithms that formally address charting delays.
LESSONS LEARNED AND THOUGHTS ON FUTURE DISSEMINATION
We believe that embedding predictive models in the EMR will become an essential component of clinical care. Despite resource limitations and having to work in a frontier area, we did 3 things well. We were able to embed a complex set of equations and display their outputs in a commercial EMR outside the research setting. In a setting where hospitalists could have requested discontinuation of the system, we achieved consensus that it should remain the standard of care. Lastly, as a result of this work, KPNC will be deploying this early warning system in all its hospitals, so our overall implementation and communication strategy has been sound.
Nonetheless, our road to implementation has been a bumpy one, and we have learned a number of valuable lessons that are being incorporated into our future work. They merit sharing with the broader medical community. Using the title of a song by Ricky SkaggsIf I Had It All Again to Dowe can summarize what we learned with 3 phrases: engage leadership early, provide simpler explanations, and embed the evaluation in the solution.
Although our research on risk adjustment and the epidemiology was known to many KPNC leaders and clinicians, our initial engagement focus was on connecting with hospital physicians and operational leaders who worked in quality improvement. In retrospect, the research team should have engaged with 2 different communities much soonerthe information technology community and that component of leadership that focused on the EMR and information technology issues. Although these 2 broad communities interact with operations all the time, they do not necessarily have regular contact with research developments that might affect both EMR as well as quality improvement operations simultaneously. Not seeking this early engagement probably slowed our work by 9 to 15 months, because of repeated delays resulting from our assumption that the information technology teams understood things that were clear to us but not to them. One major result of this at KPNC is that we now have a regular quarterly meeting between researchers and the EMR leadership. The goal of this regular meeting is to make sure that operational leaders and researchers contemplating projects with an informatics component communicate early, long before any consideration of implementation occurs.
Whereas the notion of providing early warning seems intuitive and simple, translating this into a set of equations is challenging. However, we have found that developing equations is much easier than developing communication strategies suitable for people who are not interested in statistics, a group that probably constitutes the majority of clinicians. One major result of this learning now guiding our work is that our team devotes more time to considering existing and possible workflows. This process includes spending more time engaging with clinicians around how they use information. We are also experimenting with different ways of illustrating statistical concepts (eg, probabilities, likelihood ratios).
As is discussed in the article by Dummett et al.,[20] 1 workflow component that remains unresolved is that of documentation. It is not clear what the documentation standard should be for a deterioration probability. Solving this particular conundrum is not something that can be done by electronic or statistical means. However, also with the benefit of hindsight, we now know that we should have put more energy into automated electronic tools that provide support for documentation after an alert. In addition to being requested by clinicians, having tools that automatically generate tracers as part of both the alerting and documentation process would also make evaluation easier. For example, it would permit a better delineation of the causal path between the intervention (providing a deterioration probability) and patient outcomes. In future projects, incorporation of such tools will get much more prominence.
Acknowledgements
The authors thank Dr. Michelle Caughey, Dr. Philip Madvig, Dr. Patricia Conolly, and Ms. Barbara Crawford for their administrative support, Dr. Tracy Lieu for reviewing the manuscript, and Ms. Rachel Lesser for formatting the manuscript.
Disclosures: This work was supported by a grant from the Gordon and Betty Moore Foundation (Early Detection, Prevention, and Mitigation of Impending Physiologic Deterioration in Hospitalized Patients Outside Intensive Care: Phase 3, pilot), The Permanente Medical Group, Inc., and Kaiser Foundation Hospitals, Inc. As part of our agreement with the Gordon and Betty Moore Foundation, we made a commitment to disseminate our findings in articles such as this one. However, the Foundation and its staff played no role in how we actually structured our articles, nor did they review or preapprove any of the manuscripts submitted as part of the dissemination component. Dr. Liu was supported by the National Institute for General Medical Sciences award K23GM112018. None of the sponsors had any involvement in our decision to submit this manuscript or in the determination of its contents. None of the authors has any conflicts of interest to declare of relevance to this work
Patients who deteriorate in the hospital and are transferred to the intensive care unit (ICU) have higher mortality and greater morbidity than those directly admitted from the emergency department.[1, 2, 3] Rapid response teams (RRTs) were created to address this problem.[4, 5] Quantitative tools, such as the Modified Early Warning Score (MEWS),[6] have been used to support RRTs almost since their inception. Nonetheless, work on developing scores that can serve as triggers for RRT evaluation or intervention continues. The notion that comprehensive inpatient electronic medical records (EMRs) could support RRTs (both as a source of patient data and a platform for providing alerts) has intuitive appeal. Not surprisingly, in addition to newer versions of manual scores,[7] electronic scores are now entering clinical practice. These newer systems are being tested in research institutions,[8] hospitals with advanced capabilities,[9] and as part of proprietary systems.[10] Although a fair amount of statistical information (eg, area under the receiver operator characteristic curve of a given predictive model) on the performance of various trigger systems has been published, existing reports have not described details of how the electronic architecture is integrated with clinical practice.
Electronic alert systems generated from physiology‐based predictive models do not yet constitute mature technologies. No consensus or legal mandate regarding their role yet exists. Given this situation, studying different implementation approaches and their outcomes has value. It is instructive to consider how a given institutional solution addresses common contingenciesoperational constraints that are likely to be present, albeit in different forms, in most placesto help others understand the limitations and issues they may present. In this article we describe the structure of an EMR‐based early warning system in 2 pilot hospitals at Kaiser Permanente Northern California (KPNC). In this pilot, we embedded an updated version of a previously described early warning score[11] into the EMR. We will emphasize how its components address institutional, operational, and technological constraints. Finally, we will also describe unfinished businesschanges we would like to see in a future dissemination phase. Two important aspects of the pilot (development of a clinical response arm and addressing patient preferences with respect to supportive care) are being described elsewhere in this issue of the Journal of Hospital Medicine. Analyses of the actual impact on patient outcomes will be reported elsewhere; initial results appear favorable.[12]
INITIAL CONSTRAINTS
The ability to actually prevent inpatient deteriorations may be limited,[13] and doubts regarding the value of RRTs persist.[14, 15, 16] Consequently, work that led to the pilot occurred in stages. In the first stage (prior to 2010), our team presented data to internal audiences documenting the rates and outcomes of unplanned transfers from the ward to the ICU. Concurrently, our team developed a first generation risk adjustment methodology that was published in 2008.[17] We used this methodology to show that unplanned transfers did, in fact, have elevated mortality, and that this persisted after risk adjustment.[1, 2, 3] This phase of our work coincided with KPNC's deployment of the Epic inpatient EMR (
Once we demonstrated that we could, in fact, predict inpatient deteriorations, we still had to address medicallegal considerations, the need for a clinical response arm, and how to address patient preferences with respect to supportive or palliative care. To address these concerns and ensure that the implementation would be seamlessly integrated with routine clinical practice, our team worked for 1 year with hospitalists and other clinicians at the pilot sites prior to the go‐live date.
The primary concern from a medicallegal perspective is that once results from a predictive model (which could be an alert, severity score, comorbidity score, or other probability estimate) are displayed in the chart, relevant clinical information has been changed. Thus, failure to address such an EMR item could lead to malpractice risk for individuals and/or enterprise liability for an organization. After discussing this with senior leadership, they specified that it would be permissible to go forward so long as we could document that an educational intervention was in place to make sure that clinicians understood the system and that it was linked to specific protocols approved by hospitalists.
Current predictive models, including ours, generate a probability estimate. They do not necessarily identify the etiology of a problem or what solutions ought to be considered. Consequently, our senior leadership insisted that we be able to answer clinicians' basic question: What do we do when we get an alert? The article by Dummett et al.[20] in this issue of the Journal of Hospital Medicine describes how we addressed this constraint. Lastly, not all patients can be rescued. The article by Granich et al.[21] describes how we handled the need to respect patient choices.
PROCEDURAL COMPONENTS
The Gordon and Betty Moore Foundation, which funded the pilot, only had 1 restriction (inclusion of a hospital in the Sacramento, California area). The other site was selected based on 2 initial criteria: (1) the chosen site had to be 1 of the smaller KPNC hospitals, and (2) the chosen site had to be easily accessible for the lead author (G.J.E.). The KPNC South San Francisco hospital was selected as the alpha site and the KPNC Sacramento hospital as the beta site. One of the major drivers for these decisions was that both had robust palliative care services. The Sacramento hospital is a larger hospital with a more complex caseload.
Prior to the go‐live dates (November 19, 2013 for South San Francisco and April 16, 2014 for Sacramento), the executive committees at both hospitals reviewed preliminary data and the implementation plans for the early warning system. Following these reviews, the executive committees approved the deployment. Also during this phase, in consultation with our communications departments, we adopted the name Advance Alert Monitoring (AAM) as the outward facing name for the system. We also developed recommended scripts for clinical staff to employ when approaching a patient in whom an alert had been issued (this is because the alert is calibrated so as to predict increased risk of deterioration within the next 12 hours, which means that a patient might be surprised as to why clinicians were suddenly evaluating them). Facility approvals occurred approximately 1 month prior to the go‐live date at each hospital, permitting a shadowing phase. In this phase, selected physicians were provided with probability estimates and severity scores, but these were not displayed in the EMR front end. This shadowing phase permitted clinicians to finalize the response arms' protocols that are described in the articles by Dummett et al.[20] and Granich et al.[21] We obtained approval from the KPNC Institutional Review Board for the Protection of Human Subjects for the evaluation component that is described below.
EARLY DETECTION ALGORITHMS
The early detection algorithms we employed, which are being updated periodically, were based on our previously published work.[11, 18] Even though admitting diagnoses were found to be predictive in our original model, during actual development of the real‐time data extraction algorithms, we found that diagnoses could not be obtained reliably, so we made the decision to use a single predictive equation for all patients. The core components of the AAM score equation are the above‐mentioned LAPS2 and COPS2; these are combined with other data elements (Table 1). None of the scores are proprietary, and our equations could be replicated by any entity with a comprehensive inpatient EMR. Our early detection system is calibrated using outcomes that occurred 12 hours from when the alert is issued. For prediction, it uses data from the preceding 12 months for the COPS2 and the preceding 24 to 72 hours for physiologic data.
Category | Elements Included | Comment |
---|---|---|
Demographics | Age, sex | |
Patient location | Unit indicators (eg, 3 West); also known as bed history indicators | Only patients in general medicalsurgical ward, transitional care unit, and telemetry unit are eligible. Patients in the operating room, postanesthesia recovery room, labor and delivery service, and pediatrics are ineligible. |
Health services | Admission venue | Emergency department admission or not. |
Elapsed length of stay in hospital up to the point when data are scanned | Interhospital transport is common in our integrated delivery system; this data element requires linking both unit stays as well as stays involving different hospitals. | |
Status | Care directive orders | Patients with a comfort careonly order are not eligible; all other patients (full code, partial code, and do not resuscitate) are. |
Admission status | Inpatients and patients admitted for observation status are eligible. | |
Physiologic | Vital signs, laboratory tests, neurological status checks | See online Appendices and references [6] and [15] for details on how we extract, format, and transform these variables. |
Composite indices | Generic severity of illness score | See text and description in reference [15] for details on the Laboratory‐based Acute Physiology score, version 2 and the Comorbidity Point Score, version 2. |
Longitudinal comorbidity score |
During the course of developing the real‐time extraction algorithms, we encountered a number of delays in real‐time data acquisition. These fall into 2 categories: charting delay and server delay. Charting delay is due to nonautomated charting of vital signs by nurses (eg, a nurse obtains vital signs on a patient, writes them down on paper, and then enters them later). In general, this delay was in the 15‐ to 30‐minute range, but occasionally was as high as 2 hours. Server delay, which was variable and ranged from a few minutes to (occasionally) 1 to 2 hours, is due to 2 factors. The first is that certain point of care tests were not always uploaded into the EMR immediately. This is because the testing units, which can display results to clinicians within minutes, must be physically connected to a computer for uploading results. The second is the processing time required for the system to cycle through hundreds of patient records in the context of a very large EMR system (the KPNC Epic build runs in 6 separate geographic instances, and our system runs in 2 of these). Figure 1 shows that each probability estimate thus has what we called an uncertainty period of 2 hours (the +2 hours addresses the fact that we needed to give clinicians a minimum time to respond to an alert). Given limited resources and the need to balance accuracy of the alerts, adequate lead time, the presence of an uncertainty period, and alert fatigue, we elected to issue alerts every 6 hours (with the exact timing based on facility preferences).
A summary of the components of our equation is provided in the Supporting Information, Appendices, in the online version of this article. The statistical performance characteristics of our final equation, which are based on approximately 262 million individual data points from 650,684 hospitalizations in which patients experienced 20,471 deteriorations, is being reported elsewhere. Between November 19, 2013 and November 30, 2015 (the most recent data currently available to us for analysis), a total of 26,386 patients admitted to the ward or transitional care unit at the 2 pilot sites were scored by the AAM system, and these patients generated 3,881 alerts involving a total of 1,413 patients, which meant an average of 2 alerts per day at South San Francisco and 4 alerts per day in Sacramento. Resource limitations have precluded us from conducting formal surveys to assess clinician acceptance. However, repeated meetings with both hospitalists as well as RRT nurses indicated that favorable departmental consensus exists.
INSTANTIATION OF ALGORITHMS IN THE EMR
Given the complexity of the calculations involving many variables (Table 1), we elected to employ Web services to extract data for processing using a Java application outside the EMR, which then pushed results into the EMR front end (Figure 2). Additional details on this decision are provided in the Supporting Information, Appendices, in the online version of this article. Our team had to expend considerable resources and time to map all necessary data elements in the real time environment, whose identifying characteristics are not the same as those employed by the KPHC data warehouse. Considerable debugging was required during the first 7 months of the pilot. Troubleshooting for the application was often required on very short notice (eg, when the system unexpectedly stopped issuing alerts during a weekend, or when 1 class of patients suddenly stopped receiving scores). It is likely that future efforts to embed algorithms in EMRs will experience similar difficulties, and it is wise to budget so as maximize available analytic and application programmer resources.
Figure 3 shows the final appearance of the graphical user interface at KPHC, which provides clinicians with 3 numbers: ADV ALERT SCORE (AAM score) is the probability of experiencing unplanned transfer within the next 12 hours, COPS is the COPS2, and LAPS is the LAPS2 assigned at the time a patient is placed in a hospital room. The current protocol in place is that the clinical response arm is triggered when the AAM score is 8.
LIMITATIONS
One of the limitations of working with a commercial EMR in a large system, such as KPNC, is that of scalability. Understandably, the organization is reluctant to make changes in the EMR that will not ultimately be deployed across all hospitals in the system. Thus, any significant modification of the EMR or its associated workflows must, from the outset, be structured for subsequent spread to the remaining hospitals (19 in our case). Because we had not deployed a system like this before, we did not know what to expect and, had we known then what experience has taught us, our initial requests would have been different. Table 2 summarizes the major changes we would have made to our implementation strategy had we known then what we know now.
Component | Status in Pilot Application | Desirable Changes |
---|---|---|
| ||
Degree of disaster recovery support | System outages are handled on an ad hoc basis. | Same level of support as is seen in regular clinical systems (24/7 technical support). |
Laboratory data feed | Web service. | It would be extremely valuable to have a definite answer about whether alternative data feeds would be faster and more reliable. |
LAPS2 score | Score appears only on ward or TCU patients. | Display for all hospitalized adults (include anyone 18 years and include ICU patients). |
Score appears only on inpatient physician dashboard. | Display scores in multiple dashboards (eg, emergency department dashboard). | |
COPS2 score | Score appears only on ward or TCU patients. | Display for all hospitalized adults (include anyone 18 years and include ICU patients). |
Score appears only on inpatient physician dashboard. | Display scores in multiple dashboards (eg, emergency department dashboard). | |
Alert response tracking | None is available. | Functionality that permits tracking what the status is of patients in whom an alert was issued (who responded, where it is charted, etc.)could be structured as a workbench report in KP HealthConnectvery important because of medical legal reasons. |
Trending capability for scores | None is available. | Trending display available in same location where vital signs and laboratory test results are displayed. |
Messaging capability | Not currently available. | Transmission of scores to rapid response team (or other designated first responder) via a smartphone, thus obviating the need for staff to check the inpatient dashboard manually every 6 hours. |
EVALUATION STRATEGY
Due to institutional constraints, it is not possible for us to conduct a gold standard pilot using patient‐level randomization, as described by Kollef et al.[8] Consequently, in addition to using the pilot to surface specific implementation issues, we had to develop a parallel scoring system for capturing key data points (scores, outcomes) not just at the 2 pilot sites, but also at the remaining 19 KPNC hospitals. This required that we develop electronic tools that would permit us to capture these data elements continuously, both prospectively as well as retrospectively. Thus, to give an example, we developed a macro that we call LAPS2 any time that permits us to assign a retrospective severity score given any T0. Our ultimate goal is to evaluate the system's deployment using a stepped wedge design[22] in which geographically contiguous clusters of 2 to 4 hospitals go live periodically. The silver standard (a cluster trial involving randomization at the individual hospital level[23]) is not feasible because KPNC hospitals span a very broad geographic area, and it is more resource intensive in a shorter time span. In this context, the most important output from a pilot such as this is to generate an estimate of likely impact; this estimate then becomes a critical component for power calculations for the stepped wedge.
Our ongoing evaluation has all the limitations inherent in the analysis of nonrandomized interventions. Because it only involves 2 hospitals, it is difficult to assess variation due to facility‐specific factors. Finally, because our priority was to avoid alert fatigue, the total number of patients who experience an alert is small, limiting available sample size. Given these constraints, we will employ a counterfactual method, multivariate matching,[24, 25, 26] so as to come as close as possible to simulating a randomized trial. To control for hospital‐specific factors, matching will be combined with difference‐in‐differences[27, 28] methodology. Our basic approach takes advantage of the fact that, although our alert system is currently running in 2 hospitals, it is possible for us to assign a retrospective alert to patients at all KPNC hospitals. Using multivariate matching techniques, we will then create a cohort in which each patient who received an alert is matched to 2 patients who are given a retrospective virtual alert during the same time period in control facilities. The pre‐ and postimplementation outcomes of pilot and matched controls are compared. The matching algorithms specify exact matches on membership status, whether or not the patient had been admitted to the ICU prior to the first alert, and whether or not the patient was full code at the time of an alert. Once potential matches are found using the above procedures, our algorithms seek the closest match for the following variables: age, alert probability, COPS2, and admission LAPS2. Membership status is important, because many individuals who are not covered by the Kaiser Foundation Health Plan, Inc., are hospitalized at KPNC hospitals. Because these nonmembers' postdischarge outcomes cannot be tracked, it is important to control for this variable in our analyses.
Our electronic evaluation strategy also can be used to quantify pilot effects on length of stay (total, after an alert, and ICU), rehospitalization, use of hospice, mortality, and cost. However, it is not adequate for the evaluation of whether or not patient preferences are respected. Consequently, we have also developed manual review instruments for structured electronic chart review (the coding form and manual are provided in the online Appendix of the article in this issue of Journal of Hospital Medicine by Granich et al.[21]). This review will focus on issues such as whether or not patients' surrogates were identified, whether goals of care were discussed, and so forth. In those cases where patients died in the hospital, we will also review whether death occurred after resuscitation, whether family members were present, and so forth.
As noted above and in Figure 1, charting delays can result in uncertainty periods. We have found that these delays can also result in discrepancies in which data extracted from the real time system do not match those extracted from the data warehouse. These discrepancies can complicate creation of analysis datasets, which in turn can lead to delays in completing analyses. Such delays can cause significant problems with stakeholders. In retrospect, we should have devoted more resources to ongoing electronic audits and to the development of algorithms that formally address charting delays.
LESSONS LEARNED AND THOUGHTS ON FUTURE DISSEMINATION
We believe that embedding predictive models in the EMR will become an essential component of clinical care. Despite resource limitations and having to work in a frontier area, we did 3 things well. We were able to embed a complex set of equations and display their outputs in a commercial EMR outside the research setting. In a setting where hospitalists could have requested discontinuation of the system, we achieved consensus that it should remain the standard of care. Lastly, as a result of this work, KPNC will be deploying this early warning system in all its hospitals, so our overall implementation and communication strategy has been sound.
Nonetheless, our road to implementation has been a bumpy one, and we have learned a number of valuable lessons that are being incorporated into our future work. They merit sharing with the broader medical community. Using the title of a song by Ricky SkaggsIf I Had It All Again to Dowe can summarize what we learned with 3 phrases: engage leadership early, provide simpler explanations, and embed the evaluation in the solution.
Although our research on risk adjustment and the epidemiology was known to many KPNC leaders and clinicians, our initial engagement focus was on connecting with hospital physicians and operational leaders who worked in quality improvement. In retrospect, the research team should have engaged with 2 different communities much soonerthe information technology community and that component of leadership that focused on the EMR and information technology issues. Although these 2 broad communities interact with operations all the time, they do not necessarily have regular contact with research developments that might affect both EMR as well as quality improvement operations simultaneously. Not seeking this early engagement probably slowed our work by 9 to 15 months, because of repeated delays resulting from our assumption that the information technology teams understood things that were clear to us but not to them. One major result of this at KPNC is that we now have a regular quarterly meeting between researchers and the EMR leadership. The goal of this regular meeting is to make sure that operational leaders and researchers contemplating projects with an informatics component communicate early, long before any consideration of implementation occurs.
Whereas the notion of providing early warning seems intuitive and simple, translating this into a set of equations is challenging. However, we have found that developing equations is much easier than developing communication strategies suitable for people who are not interested in statistics, a group that probably constitutes the majority of clinicians. One major result of this learning now guiding our work is that our team devotes more time to considering existing and possible workflows. This process includes spending more time engaging with clinicians around how they use information. We are also experimenting with different ways of illustrating statistical concepts (eg, probabilities, likelihood ratios).
As is discussed in the article by Dummett et al.,[20] 1 workflow component that remains unresolved is that of documentation. It is not clear what the documentation standard should be for a deterioration probability. Solving this particular conundrum is not something that can be done by electronic or statistical means. However, also with the benefit of hindsight, we now know that we should have put more energy into automated electronic tools that provide support for documentation after an alert. In addition to being requested by clinicians, having tools that automatically generate tracers as part of both the alerting and documentation process would also make evaluation easier. For example, it would permit a better delineation of the causal path between the intervention (providing a deterioration probability) and patient outcomes. In future projects, incorporation of such tools will get much more prominence.
Acknowledgements
The authors thank Dr. Michelle Caughey, Dr. Philip Madvig, Dr. Patricia Conolly, and Ms. Barbara Crawford for their administrative support, Dr. Tracy Lieu for reviewing the manuscript, and Ms. Rachel Lesser for formatting the manuscript.
Disclosures: This work was supported by a grant from the Gordon and Betty Moore Foundation (Early Detection, Prevention, and Mitigation of Impending Physiologic Deterioration in Hospitalized Patients Outside Intensive Care: Phase 3, pilot), The Permanente Medical Group, Inc., and Kaiser Foundation Hospitals, Inc. As part of our agreement with the Gordon and Betty Moore Foundation, we made a commitment to disseminate our findings in articles such as this one. However, the Foundation and its staff played no role in how we actually structured our articles, nor did they review or preapprove any of the manuscripts submitted as part of the dissemination component. Dr. Liu was supported by the National Institute for General Medical Sciences award K23GM112018. None of the sponsors had any involvement in our decision to submit this manuscript or in the determination of its contents. None of the authors has any conflicts of interest to declare of relevance to this work
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6(2):74–80. , , , , , .
- Adverse outcomes associated with delayed intensive care unit transfers in an integrated healthcare system. J Hosp Med. 2012;7(3):224–230. , , , .
- Risk factors for unplanned transfer to intensive care within 24 hours of admission from the emergency department in an integrated healthcare system. J Hosp Med. 2012;8(1):13–19. , , , , , .
- The medical emergency team: a new strategy to identify and intervene in high‐risk surgical patients. Clin Intensive Care. 1995;6:269–272. , , , , .
- The medical emergency team. Anaesth Intensive Care. 1995;23(2):183–186. , , , .
- The critically ill: following your MEWS. QJM. 2001;94(10):507–510. .
- National Health Service. National Early Warning Score (NEWS). Standardising the Assessment Of Acute‐Illness Severity in the NHS. Report of a Working Party. London, United Kingdom: Royal College of Physicians; 2012.
- A randomized trial of real‐time automated clinical deterioration alerts sent to a rapid response team. J Hosp Med. 2014;9(7):424–429. , , , et al.
- Automated detection of physiologic deterioration in hospitalized patients. J Am Med Inform Assoc. 2015;22(2):350–360. , , , et al.
- Identifying patients at increased risk for unplanned readmission. Med Care. 2013;51(9):761–766. , , , , .
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- Early detection of impending deterioration outside the ICU: a difference‐in‐differences (DiD) study. Presented at: American Thoracic Society International Conference, San Francisco, California; May 13–18, 2016; A7614. , , , et al.
- Unplanned transfers to a medical intensive care unit: causes and relationship to preventable errors in care. J Hosp Med. 2011;6(2):68–72. , , , .
- Rapid response teams—walk, don't run. JAMA. 2006;296(13):1645–1647. , , .
- Rapid response systems: a systematic review. Crit Care Med. 2007;35(5):1238–1243. , , , , , .
- Rethinking rapid response teams. JAMA. 2010;304(12):1375–1376. , .
- Risk adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated healthcare delivery system. Med Care. 2013;51(5):446–453. , , , , .
- Nonelective rehospitalizations and post‐discharge mortality: predictive models suitable for use in real time. Med Care. 2015;53(11):916–923. , , , , , .
- J Hosp Med. 2016;11:000–000. et al.
- J Hosp Med. 2016;11:000–000. et al.
- Design and analysis of stepped wedge cluster randomized trials. Contemp Clin Trials. 2007;28(2):182–191. , .
- Cluster randomized trials: evaluating treatments applied to groups. JAMA. 2015;313(20):2068–2069. , .
- Comparison of multivariate matching methods: structures, distances, and algorithms. J Comput Graph Stat. 1993;2(4):405–420. , .
- A method/macro based on propensity score and Mahalanobis distance to reduce bias in treatment comparison in observational study. Eli Lilly working paper available at: http://www.lexjansen.com/pharmasug/2006/publichealthresearch/pr05.pdf. , , .
- Matching methods for causal inference: a review and a look forward. Stat Sci. 2010;25(1):1–21. .
- Methods for evaluating changes in health care policy: the difference‐in‐differences approach. JAMA. 2014;312(22):2401–2402. , .
- Why we should not be indifferent to specification choices for difference‐in‐differences. Health Serv Res. 2015;50(4):1211–1235. , , .
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6(2):74–80. , , , , , .
- Adverse outcomes associated with delayed intensive care unit transfers in an integrated healthcare system. J Hosp Med. 2012;7(3):224–230. , , , .
- Risk factors for unplanned transfer to intensive care within 24 hours of admission from the emergency department in an integrated healthcare system. J Hosp Med. 2012;8(1):13–19. , , , , , .
- The medical emergency team: a new strategy to identify and intervene in high‐risk surgical patients. Clin Intensive Care. 1995;6:269–272. , , , , .
- The medical emergency team. Anaesth Intensive Care. 1995;23(2):183–186. , , , .
- The critically ill: following your MEWS. QJM. 2001;94(10):507–510. .
- National Health Service. National Early Warning Score (NEWS). Standardising the Assessment Of Acute‐Illness Severity in the NHS. Report of a Working Party. London, United Kingdom: Royal College of Physicians; 2012.
- A randomized trial of real‐time automated clinical deterioration alerts sent to a rapid response team. J Hosp Med. 2014;9(7):424–429. , , , et al.
- Automated detection of physiologic deterioration in hospitalized patients. J Am Med Inform Assoc. 2015;22(2):350–360. , , , et al.
- Identifying patients at increased risk for unplanned readmission. Med Care. 2013;51(9):761–766. , , , , .
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- Early detection of impending deterioration outside the ICU: a difference‐in‐differences (DiD) study. Presented at: American Thoracic Society International Conference, San Francisco, California; May 13–18, 2016; A7614. , , , et al.
- Unplanned transfers to a medical intensive care unit: causes and relationship to preventable errors in care. J Hosp Med. 2011;6(2):68–72. , , , .
- Rapid response teams—walk, don't run. JAMA. 2006;296(13):1645–1647. , , .
- Rapid response systems: a systematic review. Crit Care Med. 2007;35(5):1238–1243. , , , , , .
- Rethinking rapid response teams. JAMA. 2010;304(12):1375–1376. , .
- Risk adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated healthcare delivery system. Med Care. 2013;51(5):446–453. , , , , .
- Nonelective rehospitalizations and post‐discharge mortality: predictive models suitable for use in real time. Med Care. 2015;53(11):916–923. , , , , , .
- J Hosp Med. 2016;11:000–000. et al.
- J Hosp Med. 2016;11:000–000. et al.
- Design and analysis of stepped wedge cluster randomized trials. Contemp Clin Trials. 2007;28(2):182–191. , .
- Cluster randomized trials: evaluating treatments applied to groups. JAMA. 2015;313(20):2068–2069. , .
- Comparison of multivariate matching methods: structures, distances, and algorithms. J Comput Graph Stat. 1993;2(4):405–420. , .
- A method/macro based on propensity score and Mahalanobis distance to reduce bias in treatment comparison in observational study. Eli Lilly working paper available at: http://www.lexjansen.com/pharmasug/2006/publichealthresearch/pr05.pdf. , , .
- Matching methods for causal inference: a review and a look forward. Stat Sci. 2010;25(1):1–21. .
- Methods for evaluating changes in health care policy: the difference‐in‐differences approach. JAMA. 2014;312(22):2401–2402. , .
- Why we should not be indifferent to specification choices for difference‐in‐differences. Health Serv Res. 2015;50(4):1211–1235. , , .
© 2016 Society of Hospital Medicine
The Learning Hospital System
In the landmark Best Care at Lower Cost report, the Institute of Medicine presents a compelling vision of a US healthcare system where science, information technology, incentives, and care culture are brought together seamlessly to produce high‐quality healthcare.[1] At the center of this transformation is the learning healthcare system, a system characterized by its ability to leverage data arising from care provision to drive rapid improvements in care delivery.[2] When steeped within the right organizational milieu, these data help to close the virtuous cycle of continuous learning moving from science to evidence to care and back to new science. The anticipated end result is a healthcare system that can provide Americans with superior care at lower cost.
Hospital‐based practitioners will recognize the inpatient setting as an ideal demonstration opportunity for continuous learning. Hospital care is costly, accounting for more than 30% of all US healthcare costs[3]; intensive care alone accounts for a notable proportion of the US gross domestic product.[4] Inpatient care is associated with significant mortality and morbidity, and its use is often greatly increased in patients' last days.[5, 6] Fortunately, the inpatient setting also offers an ideal opportunity to leverage high‐quality data to help inform and improve care. The digitization of medicine means that far more data are now available through electronic health records, medical devices, and tests.[7] This is particularly true for inpatients, for whom a large volume of data are produced even over relatively short hospital stays.
Whereas the challenge to improve hospital care is daunting, there is an incredible opportunity to advance the quality of inpatient care through realizing the vision of the learning hospital system. In the sections that follow, we use an object lessonsepsis care within hospitals of the Kaiser Permanente Northern California (KPNC) integrated healthcare delivery systemto evaluate the challenges and insights gleaned from working toward building a learning hospital system. Then, we describe further steps that could enhance the use of inpatient data to drive improved care.
THE FRAMEWORK OF A LEARNING HEALTHCARE SYSTEM
Best Care at Lower Cost notes a fundamental paradox in US healthcare: although we have witnessed a dramatic expansion in biomedical knowledge, innovative therapies and surgical procedures, and clinical treatments to extend survival, US healthcare persistently falls short on the basic dimensions of quality, outcomes, cost, and equity.[1] The proposed path forward lies in building the learning healthcare system, a system characterized by continuous knowledge development, improvement, and application. Figure 1 shows the critical nodes in the framework for continuous learning, which include: (1) the development of new scientific knowledge (science), (2) the translation of science into clinical evidence of efficacy (evidence), and (3) the application of efficacious interventions through effective care delivery (care). In healthcare today, transitions between these nodes are rife with missed or wasted opportunities like delays in applying high‐quality evidence or poorly managed insights arising from scientific discovery. If such opportunities could be recovered, however, the quality of healthcare could be improved dramatically.[8]
The pursuit of continuous learning is aided by rapid changes in the quality and quantity of biomedical data available over the past decade, especially through the use of electronic health records, novel biomolecular tools, and digital sensors.[2, 7, 9] The Internet has ushered in a new era of data connectivity, for example, allowing for highly engaged communication between patients and providers as well as collaboration between professional or citizen scientists on data of unprecedented scale.[10] New methodologic approaches, including data mining and machine learning, increasingly leverage commodity hardware to conduct previously computationally intractable analyses.[9] Moreover, the development of domain ontologies fosters the discovery of meaningful insights from data of heterogeneous types.[11]
Ultimately, however, improvements in data alone are inadequate to achieve continuous learning. As shown in Figure 1, whereas data form the channels that allow for transitions from science to evidence to care, novel insights need to be steeped within the right culture, motivated by the right incentives, and supported by the right leaders.[1, 12] Within the sustainable learning healthcare system, knowledge generation feeds practice change with the support and guidance of system leadership; improved practice, in turn, generates new knowledge and completes the virtuous cycle of learning.
THE PROMISE OF CONTINUOUS LEARNING IN HOSPITAL SETTINGS
The hospital is an ideal setting in which to foster continuous learning because advances in inpatient care have the potential to substantially improve healthcare quality and value.[8] Americans were hospitalized roughly 37 million times in 2012; in total, these episodes cost $378 billion.[3] Over 700,000 patients die in US hospitals annually, with reports showing that many patients utilize greatly increased inpatient and critical care services near the end of their lives in a manner that appears misaligned with their preferences.[11, 13] Hospital care is also highly variable in quality and cost; this heterogeneity is not closely associated with improved outcomes.[14, 15] Preventable harm and medical injury occur commonly in hospitals and are now recognized to be a leading cause of inpatient death.[16] Finally, emerging research illuminates the substantial toll that acute care has on patients and families resulting in new comorbidity, functional or neuropsychiatric impairment, rehospitalization, and financial burden that persist long after patients are discharged.[17]
Fortunately, inpatient care also exhibits several qualities that improve the likelihood that continuous learning can be achieved. Although it is clear that hospitalizations occur within the arc of a patient's larger health trajectory, these distinct episodes offer the potential to observe patient trajectories and treatments evolving within relatively compressed time intervals; over that same interval, a large volume of data are produced. Stored within comprehensive electronic health records, these granular data now allow inpatient episodes to be digitally recapitulated with high fidelity, bolstering their use in driving care improvements.[18]
AN OBJECT LESSON IN THE LEARNING FRAMEWORK: SEPSIS CARE
Translating Science to Evidence in Sepsis
Although sepsis has attracted great attention in modern hospital care, sepsis was described long ago by Hippocrates to describe the process by which wounds fester.[19] Recast after the confirmation of germ theory, sepsis came to be known primarily as the blood poisoning resulting from pathogenic organisms.[20] However, with the advent of antibiotics, numerous scientific studies now recognize that sepsis actually results from the dysregulated host immune response to systemic infection, which can also cause organ dysfunction.[21] Based on this knowledge, landmark translational and clinical studies in the 2000s provided strong evidence that early identification of sepsis patients and aggressive infection control and resuscitation were associated with improved mortality (Figure 2, step 1).[22]
Translating Evidence to Care in Sepsis at KPNC
In 2007, the leadership of KPNC initiated a regional effort to improve the quality of care and reduce the variability in performance at its medical centers (Table 1).[23] Reviewing data from nearly 1000 inpatientsthe last 50 consecutive hospital deaths from each of 19 medical centersa mortality diagnostic based on Institute for Healthcare Improvement recommendations[24] revealed that sepsis had a major impact on hospital outcomes. For example, even though sepsis patients were still relatively under‐recognized at the time, accounting for fewer than 3% of hospitalizations, they contributed to one‐quarter of hospital deaths. In light of these compelling data, senior regional leadership identified reducing sepsis mortality as a key performance improvement goal (Figure 2, step 2).
Time Period | Event Summary |
---|---|
| |
2007 | Variability in hospital standardized mortality ratio observed, indicating an opportunity to drive improved outcomes. |
Initiation of staggered implementation of unified electronic medical record across all KP sites (starting in 2006 and ending in 2009). | |
Spring 2008 | Mortality diagnostic chart review completed identifying sepsis and infection‐related causes as key factors in hospital outcomes. |
May 2008 | Regional Mortality Summit held with a focus on patient safety and mortality reduction efforts through performance improvement. Executive regional and local leadership alignment to focus on sepsis performance improvement. |
Summer 2008 | Sepsis Steering Committee evaluates best available evidence, develops treatment algorithms, and plans for medical center pilots. |
Fall 2008 | Pilot intervention deployed at 2 medical centers. |
November 2008 | First Regional Sepsis Summit: development of sepsis performance improvement playbook, training materials, implementation plans, and measurement strategy. |
November 2008 | All medical centers begin to form multidisciplinary sepsis teams and performance improvement committees, obtain equipment and supplies including assembly of a sepsis cart. Multidisciplinary teams included ED physician champion, ED nurse champion, improvement advisor, hospitalists, intensivists, quality improvement personnel, nurse educators, and even resident physicians. |
January 2009 | Performance data collection begins on EGDT processes and outcomes. Initiation of 2 key elements to enhance screening for and detection of sepsis: (1) concomitant ordering of serum lactic acid along with blood cultures, and (2) definition of lactate >2.0 as a critical lab value. |
Use of manual chart review for case finding and central database entry because of ongoing implementation of electronic medical record and limited sepsis‐specific data infrastructure. | |
March 2009 | Regional train the trainer sessions occur and local educational spread efforts begin including: collaborative calls, in‐person training events, and medical center site visits. |
August 2009 | Grant funding from the Gordon and Betty Moore Foundation begins with a planned 2‐year duration providing funding for improvement advisors with performance improvement expertise and data infrastructure development. |
November 2009 | Second Regional Sepsis Summit. Identification of intermediate lactate sepsis patients having significant mortality. |
January 2010 | Initiate measurement of performance for intermediate lactate sepsis patients with a focus on lactate clearance as an outcome measure of interest. |
2010 | Development of an intranet Web‐based data abstraction tool to identify cases and auto‐populate specific fields for review. Facilities were responsible for review of cases at the local level to foster rapid feedback cycles for local performance improvement. Standardized data query tools were deployed to foster local medical center engagement and system‐level evaluation. |
Accompanying development of a sepsis performance improvement scorecard allowing for comparison of longitudinal performance metrics across all facilities. Scorecard elements included: proportion of lactates drawn following ED blood culture, EGDT‐specific bundle elements (ie, number of EGDT cases, antibiotics within 1 hour, first central venous pressure within 2 hours of EGDT start, target mean arterial pressure achievement), repeat lactate elements, balancing measures for central line placement (ie, pneumothorax, central line infection), and overall sepsis statistics. | |
April 2011 | Third Regional Sepsis Summit. Refinement of EGDT bundle and further development of intermediate lactate bundle approach, including piloting specific treatment bundles targeting this population. Collaborative performance improvement environment in which successful strategies at 1 site were rapidly disseminated to other sites including the Sepsis Alert and the Sepsis Clock. |
May 2012 | Research analysis of fluid volume and lactate clearance in intermediate lactate sepsis population begins. |
February 2013 | Fourth Regional Sepsis Summit. Regional spread of intermediate lactate bundle including the use of fluids, antibiotics, and repeat lactate measurements. |
May 2013 | Research analysis of the contribution of sepsis to hospital deaths (within KP and in national sample) as well as post‐sepsis resource utilization and mortality |
March 2014 | Publication of ProCESS randomized clinical trial, requiring systemic reevaluation of EGDT‐based sepsis strategy. Subsequent publications of ARISE and ProMISe trials confirming findings from ProCESS. Updated approach under consideration and informally disseminated to practitioners. |
October 2014 | Updated sepsis treatment guidelines and data capture strategy fully implemented moving away from a catheter‐based strategy for all EGDT‐eligible patients. |
October 2015 | Sixth Regional Sepsis Summit held to adjust sepsis treatment and data measurement strategy to align more closely with CMS SEP‐1 guidelines. |
Based on the principles of performance improvement methodology, clinical and operational leaders established an environment with aligned culture, incentives, and leadership around sepsis care. The effort was launched in late 2008 at a Sepsis Summit, bringing together a multidisciplinary group of stakeholders (eg, hospitalist, emergency department, and intensive care chiefs of staff and nursing managers; medical center and nursing executive and operational leadership) and providing sepsis care pathways based on the best available evidence.[23] Regional investments in the digital infrastructure to support implementation resulted in the provision of granular data within monthly sepsis scorecards quantifying each medical center's performance and trends for a diverse set of sepsis bundle metrics.
The resulting changes in sepsis care were substantial. For example, improved early recognition of infected patients meeting the criteria for sepsis resulted in large changes in the standardized diagnostic criteria used to label patients (Figure 3A). Implementing screening strategies using serum lactate testing for any patient receiving blood cultures resulted in a roughly 10‐fold increase in the use of lactate testing in the emergency department (Figure 3B). Earlier recognition of sepsis also increased the number of patients receiving early antibiotics and receiving central venous catheters for quantitative resuscitation.[23]
CLOSING THE LOOP TOWARD CONTINUOUS LEARNING IN SEPSIS
Leveraging timely and actionable data steeped within an aligned organizational milieu resulted in large‐scale changes across a heterogeneous set of hospitals. However, to realize the true vision of a learning hospital system, a looming question remained: Could the data generated as the byproduct of routine care now be used to complete the virtuous cycle and drive new scientific discovery (Figure 2, step 3)?
Confirming Concordance in the Impact of Sepsis Nationally
The heightened identification of sepsis patients through program implementation revealed that the impact of sepsis on hospital mortality was greater than originally estimated; based on improved patient identification, sepsis now accounted for upward of 1 in 2 hospital deaths.[25] This sobering statistic confirmed that the investments in standardizing best sepsis care following the mortality diagnostic were critical. However, were similar estimates of sepsis‐attributable mortality consistent outside of the KPNC system? To study this, we examined US hospitalizations occurring across >1000 hospitals and involving >6 million hospital stays to estimate corresponding prevalence.[25] In this national sample, sepsis contributed to as many as half of hospital deaths in the United States in 2010, lending strong support to ongoing international and state‐based efforts to improve sepsis care. These studies also paved the way to use these data drawn from our large sepsis population to inform updated international consensus definitions for sepsis and septic shock.[26, 27, 28]
Identifying New Avenues for Reducing the Toll of Sepsis
A major concern of sepsis program leaders was the prior findings that sepsis hospitalizations among Medicare beneficiaries were associated with substantial new cognitive and functional disability.[29] This lingering toll of sepsis had been termed a hidden public health disaster.[30] To further understand the posthospital impact of sepsis and to begin investigating new avenues to reduce this impact, a cohort of patients was followed for 1 year following sepsis hospitalization.[31] Over that period, nearly half of sepsis survivors were rehospitalized. When compared with their presepsis levels of healthcare utilization, middle‐aged and elderly sepsis patients experienced a 3‐fold increase in their days using facility‐based care. Subsequent studies in other populations outside of KPNC have confirmed these concerning findings, resulting in new efforts to address postsepsis survivorship care.[32, 33]
Phenotyping New Targets for Standardized Sepsis Care
At its outset, the sepsis improvement program applied the best available evidence to treat patients with the most severe forms of sepsisseptic shock. However, once the initial implementation phase had succeeded, clinicians and operational leaders quickly realized from the emerging data that there was a far larger group of sepsis patients for whom treatment guidelines were poorly defined.[25, 34, 35] These were severe sepsis patients with so‐called intermediate lactate values between 2 mmol/L and 4 mmol/L; they comprised a substantial proportion of all sepsis patients dying in the hospital. Using data generated from the routine care of sepsis patients treated across 21 hospitals, the sepsis leadership group was able to rapidly assemble a cohort of intermediate lactate sepsis patients up to 20‐ to 100‐fold larger than that reported in prior studies and evaluate their outcomes.[34, 35]
The data used to evaluate these intermediate lactate sepsis patients now spurred a new implementation program in 2013 for a group of patients in whom there was essentially no existing evidence to guide care. Rapidly implemented within a mature sepsis performance improvement program, evaluations at the 6‐month and 1‐year intervals demonstrated significant decreases in mortality.[36] Importantly, to allay the justified concerns of clinicians, these evaluations also clearly showed no evidence of harm from more aggressive fluid resuscitation (eg, increased transfer to intensive care, increased rates of mechanical ventilation). Again, driven by clinician input, subgroup analyses further revealed that the implementation program was only associated with reduced mortality in patients who could be at risk for iatrogenic fluid overload (ie, those with a history of congestive heart failure or chronic kidney disease).[36] Spurred by these provocative findings, operational and clinical leaders are currently considering how to guide future care in these patients, especially with the emerging use of noninvasive methods to quantify patients' fluid responsiveness.
PRINCIPLES FOR LEVERAGING DATA IN THE LEARNING HOSPITAL SYSTEM
The object lesson of using data to drive improved sepsis care and further new scientific discovery offers some important insights for continuous learning.
Building a Digital Infrastructure for Utilizing Granular Hospital Data
As described above, current transitions between the nodes of the learning framework are rife with missed opportunities. Perhaps one of the most glaring is the inability to use highly granular data already collected within the electronic health record (eg, trajectories and trends across vital signs or laboratory results, large‐scale medication administration records to evaluate multidrug interactions). An essential starting point for continuous learning is investing in the digital infrastructure to improve the use of data beyond traditional claims (administrative dataadmission source codes, disposition codes, diagnoses, and procedures). As shown in Table 2, the first key step is incorporating laboratory data into the quality assessment/emmprovement process. In addition, using these data to automate severity of illness and risk adjustment metrics fosters use of similar comparison cohorts across time or disease types.[18, 37, 38, 39, 40]
Data Type | Contents | Degree of Difficulty in Accessing | Degree of Difficulty in Analyzing |
---|---|---|---|
Administrative | Traditional claims data, diagnostic or procedural codes | Low | Low to moderate |
Standard cohort profiling | Limited instances of vitals signs, laboratory, diagnostic testing, or treatment data | Low to moderate | Low to moderate |
Metrics reporting for care improvement | Standard cohort identification, aggregated achievement of treatment targets, scorecard dissemination | Moderate | Moderate |
Advanced cohort profiling | Time series of physiologic data, inpatient triage and treatment data within short temporal intervals | Moderate to high | High |
Research‐grade discovery | Data with breadth (representative sample size) and depth (highly granular physiologic and treatment data) | High | Very high |
Patient‐reported outcomes | Quality of life, functional and cognitive disability | Very high | High |
Employing Novel Methods to Address the Limitations of Using Real‐World Data
The rapid digitization of medicine through the use of electronic medical records offers tremendous opportunities to facilitate continuous learning. However, these opportunities are accompanied by important limitations.[41] Data collected as a byproduct of real‐world care can be vulnerable to many forms of bias and confounding, potentially clouding the validity and robustness of corresponding analytic results. Fortunately, advanced methods including causal inference are now used routinely to address some limitations.[42] In the context of a learning healthcare system, other opportunities for improved study design including cluster randomized trials or stepped wedge implementation can also be employed to preserve the statistical rigor of subsequent analyses.[43] Finally, emerging methods employing randomization through the electronic medical record alongside adaptive trial design offer great potential to increase the efficiency of continuous learning.[44]
Evaluating the Hospital as a Single System
Advances in contemporary hospital care require seamless transitions of patient care, screening strategies, and therapeutic approaches across multiple hospital domains and with diverse providers; these interventions also need to happen rapidly. Many traditional approaches to inpatient care have taken a bottom‐up approach (eg, studying a specific disease within a specific hospital ward like the intensive care unit) that have proven useful but may limit generalizability when applied to a real‐world hospital operating with Pareto optimality (ie, the trade‐off scenario where new resource allocation to 1 area also requires resource withdrawal from another area). In certain cases, an empiric approach, without initial preference for any specific ward or disease, can aid decision making by hospital operational and clinical leaders by providing a global picture of impact and value.
Focusing on Early Detection in Hospital Settings as Secondary Prevention
Once patients have been admitted to the hospital, a race against the clock begins. Each additional hour of hospitalization increases the risks of iatrogenic injury or medical harm manifested by immobility, disorientation and delirium, nosocomial infections, or medication errors, among others. In this context, detection systems that use granular hospital data to focus on the earliest detection of risk can aid critical approaches to secondary prevention (Although the hospitalization for sepsis cannot be avoided, careful attention to mobility can limit the risk of developing delirium. In turn, preventing delirium can limit the risk of new functional disability).
Contextualizing Hospital Care Within a Longitudinal Trajectory
Although we described the benefit of hospital episodes having well‐demarcated beginning and ending points, it remains essential to recognize that the harms associated with hospitalization extend well beyond discharge. In this context, hospitalizations can serve as waypoints in patients' health trajectories as well as an opportunity to achieve patient‐centered care including discussing and aligning goals of care with actual care provision. Furthermore, although we have seen steady declines in hospital mortality over time, it is highly likely that we will reach a nadir in mortality where additional metrics of hospital outcomes will need to include postdischarge events like readmission, long‐term mortality, quality of life, and the prevention of disability or decline.
CONCLUSION
Hospitalizations in the United States are costly and associated with high mortality and morbidity; the toll of hospitalization also extends well beyond hospital discharge. The promise of the learning hospital system has marked improvements in the quality of hospital care, especially where healthcare systems can steep critical investments in data and digital infrastructure within the right culture, incentives, and leadership. Where continuous learning is achieved, data generated during routine care offer the potential to yield new scientific discovery and drive further improvements in hospital care.
Disclosures
As part of our agreement with the Gordon and Betty Moore Foundation, we made a commitment to disseminate our findings in articles such as this one. However, the Foundation and its staff played no role in how we actually structured our articles, nor did they review or preapprove any of the manuscripts submitted as part of the dissemination component. None of the authors has any conflicts of interest to declare of relevance to this work, which was funded by a combination of funding from the Gordon and Betty Moore Foundation, The Permanente Medical Group, Inc., and Kaiser Foundation Hospitals, Inc. VXL was supported by NIH K23GM112018.
- Institute of Medicine. Best Care at Lower Cost: The Path to Continuously Learning Health Care in America. Washington, DC: The National Academies Press; 2012.
- Toward a science of learning systems: a research agenda for the high‐functioning Learning Health System. J Am Med Inform Assoc. 2015;22(1):43–50. , , , et al.
- National Center for Health Statistics. Health, United States, 2014: With Special Feature on Adults Aged 55–64. Hyattsville, MD; 2015.
- Critical care medicine in the United States 2000‐2005: an analysis of bed numbers, occupancy rates, payer mix, and costs. Crit Care Med. 2010;38(1):65–71. , .
- Trends and variation in end‐of‐life care for medicare beneficiaries with severe chronic illness. A report of the Dartmouth Atlas Project. Lebanon, NH: The Dartmouth Institute for Health Policy and Clinical Practice; 2011. , , , .
- Change in end‐of‐life care for Medicare beneficiaries: site of death, place of care, and health care transitions in 2000, 2005, and 2009. JAMA. 2013;309(5):470–477. , , , et al.
- Finding the missing link for big biomedical data. JAMA. 2014;311(24):2479–2480. , , .
- Code red and blue—safely limiting health care's GDP footprint. N Engl J Med. 2013;368(1):1–3. .
- The inevitable application of big data to health care. JAMA. 2013;309(13):1351–1352. , .
- What is citizen science?—a scientometric meta‐analysis. PLoS One. 2016;11(1):e0147152. , .
- Biomedical ontologies: a functional perspective. Brief Bioinform. 2008;9(1):75–90. , , .
- Rapid learning: a breakthrough agenda. Health Aff (Millwood). 2014;33(7):1155–1162. .
- Are regional variations in end‐of‐life care intensity explained by patient preferences?: a study of the US Medicare population. Med Care. 2007;45(5):386–393. , , , et al.
- Extreme markup: the fifty US hospitals with the highest charge‐to‐cost ratios. Health Aff (Millwood). 2015;34(6):922–928. , .
- The price ain't right? Hospital prices and health spending on the privately insured. Health Care Pricing Project website. Available at: http://www.healthcarepricingproject.org/sites/default/files/pricing_variation_manuscript_0.pdf. Accessed February 15, 2016 , , , .
- A new, evidence‐based estimate of patient harms associated with hospital care. J Patient Saf. 2013;9(3):122–128. .
- Hospitalization‐associated disability: “she was probably able to ambulate, but I'm not sure”. JAMA. 2011;306(16):1782–1793. , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated health care delivery system. Med Care. 2013;51(5):446–453. , , , , .
- New definitions for sepsis and septic shock: continuing evolution but with much still to be done. JAMA. 2016;315(8):757–759. .
- Severe sepsis and septic shock. N Engl J Med. 2013;369(21):2063. , .
- Sepsis‐induced immunosuppression: from cellular dysfunctions to immunotherapy. Nat Rev Immunol. 2013;13(12):862–874. , , .
- Early goal‐directed therapy in the treatment of severe sepsis and septic shock. N Engl J Med. 2001;345(19):1368–1377. , , , et al.
- Kaiser Permanente's performance improvement system, part 3: multisite improvements in care for patients with sepsis. Jt Comm J Qual Patient Saf. 2011;37(11):483–493. , , , et al.
- Understanding the components of quality improvement collaboratives: a systematic literature review. Milbank Q. 2013;91(2):354–394. , , , , .
- Hospital deaths in patients with sepsis from 2 independent cohorts. JAMA. 2014;312(1):90–92. , , , et al.
- Assessment of clinical criteria for sepsis: for the Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):762–774. , , , et al.
- Developing a new definition and assessing new clinical criteria for septic shock: for the Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):775–787. , , , et al.
- The Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):801–810. , , , et al.
- Long‐term cognitive impairment and functional disability among survivors of severe sepsis. JAMA. 2010;304(16):1787–1794. , , , .
- The lingering consequences of sepsis: a hidden public health disaster? JAMA. 2010;304(16):1833–1834. .
- Hospital readmission and healthcare utilization following sepsis in community settings. J Hosp Med. 2014;9(8):502–507. , , , , , .
- Increased 1‐year healthcare use in survivors of severe sepsis. Am J Respir Crit Care Med. 2014;190(1):62–69. , , , , .
- Post‐acute care use and hospital readmission after sepsis. Ann Am Thorac Soc. 2015;12(6):904–913. , , , et al.
- Fluid volume, lactate values, and mortality in sepsis patients with intermediate lactate values. Ann Am Thorac Soc. 2013;10(5):466–473. , , , , .
- Prognosis of emergency department patients with suspected infection and intermediate lactate levels: a systematic review. J Crit Care. 2014;29(3):334–339. , , .
- Multicenter implementation of a treatment bundle for sepsis patients with intermediate lactate values. Am J Respir Crit Care Med. 2016;193(11):1264–1270. , , , et al.
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases. Am J Manag Care. 2008;14(3):158–166. , , , et al.
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6(2):74–80. , , , , , .
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- An electronic Simplified Acute Physiology Score‐based risk adjustment score for critical illness in an integrated healthcare system. Crit Care Med. 2013;41(1):41–48. , , , , .
- Learning from big health care data. N Engl J Med. 2014;370(23):2161–2163. .
- Getting the methods right—the foundation of patient‐centered outcomes research. N Engl J Med. 2012;367(9):787–790. , .
- The stepped wedge cluster randomised trial: rationale, design, analysis, and reporting. BMJ. 2015;350:h391. , , , , .
- Fusing randomized trials with big data: the key to self‐learning health care systems? JAMA. 2015;314(8):767–768. .
In the landmark Best Care at Lower Cost report, the Institute of Medicine presents a compelling vision of a US healthcare system where science, information technology, incentives, and care culture are brought together seamlessly to produce high‐quality healthcare.[1] At the center of this transformation is the learning healthcare system, a system characterized by its ability to leverage data arising from care provision to drive rapid improvements in care delivery.[2] When steeped within the right organizational milieu, these data help to close the virtuous cycle of continuous learning moving from science to evidence to care and back to new science. The anticipated end result is a healthcare system that can provide Americans with superior care at lower cost.
Hospital‐based practitioners will recognize the inpatient setting as an ideal demonstration opportunity for continuous learning. Hospital care is costly, accounting for more than 30% of all US healthcare costs[3]; intensive care alone accounts for a notable proportion of the US gross domestic product.[4] Inpatient care is associated with significant mortality and morbidity, and its use is often greatly increased in patients' last days.[5, 6] Fortunately, the inpatient setting also offers an ideal opportunity to leverage high‐quality data to help inform and improve care. The digitization of medicine means that far more data are now available through electronic health records, medical devices, and tests.[7] This is particularly true for inpatients, for whom a large volume of data are produced even over relatively short hospital stays.
Whereas the challenge to improve hospital care is daunting, there is an incredible opportunity to advance the quality of inpatient care through realizing the vision of the learning hospital system. In the sections that follow, we use an object lessonsepsis care within hospitals of the Kaiser Permanente Northern California (KPNC) integrated healthcare delivery systemto evaluate the challenges and insights gleaned from working toward building a learning hospital system. Then, we describe further steps that could enhance the use of inpatient data to drive improved care.
THE FRAMEWORK OF A LEARNING HEALTHCARE SYSTEM
Best Care at Lower Cost notes a fundamental paradox in US healthcare: although we have witnessed a dramatic expansion in biomedical knowledge, innovative therapies and surgical procedures, and clinical treatments to extend survival, US healthcare persistently falls short on the basic dimensions of quality, outcomes, cost, and equity.[1] The proposed path forward lies in building the learning healthcare system, a system characterized by continuous knowledge development, improvement, and application. Figure 1 shows the critical nodes in the framework for continuous learning, which include: (1) the development of new scientific knowledge (science), (2) the translation of science into clinical evidence of efficacy (evidence), and (3) the application of efficacious interventions through effective care delivery (care). In healthcare today, transitions between these nodes are rife with missed or wasted opportunities like delays in applying high‐quality evidence or poorly managed insights arising from scientific discovery. If such opportunities could be recovered, however, the quality of healthcare could be improved dramatically.[8]
The pursuit of continuous learning is aided by rapid changes in the quality and quantity of biomedical data available over the past decade, especially through the use of electronic health records, novel biomolecular tools, and digital sensors.[2, 7, 9] The Internet has ushered in a new era of data connectivity, for example, allowing for highly engaged communication between patients and providers as well as collaboration between professional or citizen scientists on data of unprecedented scale.[10] New methodologic approaches, including data mining and machine learning, increasingly leverage commodity hardware to conduct previously computationally intractable analyses.[9] Moreover, the development of domain ontologies fosters the discovery of meaningful insights from data of heterogeneous types.[11]
Ultimately, however, improvements in data alone are inadequate to achieve continuous learning. As shown in Figure 1, whereas data form the channels that allow for transitions from science to evidence to care, novel insights need to be steeped within the right culture, motivated by the right incentives, and supported by the right leaders.[1, 12] Within the sustainable learning healthcare system, knowledge generation feeds practice change with the support and guidance of system leadership; improved practice, in turn, generates new knowledge and completes the virtuous cycle of learning.
THE PROMISE OF CONTINUOUS LEARNING IN HOSPITAL SETTINGS
The hospital is an ideal setting in which to foster continuous learning because advances in inpatient care have the potential to substantially improve healthcare quality and value.[8] Americans were hospitalized roughly 37 million times in 2012; in total, these episodes cost $378 billion.[3] Over 700,000 patients die in US hospitals annually, with reports showing that many patients utilize greatly increased inpatient and critical care services near the end of their lives in a manner that appears misaligned with their preferences.[11, 13] Hospital care is also highly variable in quality and cost; this heterogeneity is not closely associated with improved outcomes.[14, 15] Preventable harm and medical injury occur commonly in hospitals and are now recognized to be a leading cause of inpatient death.[16] Finally, emerging research illuminates the substantial toll that acute care has on patients and families resulting in new comorbidity, functional or neuropsychiatric impairment, rehospitalization, and financial burden that persist long after patients are discharged.[17]
Fortunately, inpatient care also exhibits several qualities that improve the likelihood that continuous learning can be achieved. Although it is clear that hospitalizations occur within the arc of a patient's larger health trajectory, these distinct episodes offer the potential to observe patient trajectories and treatments evolving within relatively compressed time intervals; over that same interval, a large volume of data are produced. Stored within comprehensive electronic health records, these granular data now allow inpatient episodes to be digitally recapitulated with high fidelity, bolstering their use in driving care improvements.[18]
AN OBJECT LESSON IN THE LEARNING FRAMEWORK: SEPSIS CARE
Translating Science to Evidence in Sepsis
Although sepsis has attracted great attention in modern hospital care, sepsis was described long ago by Hippocrates to describe the process by which wounds fester.[19] Recast after the confirmation of germ theory, sepsis came to be known primarily as the blood poisoning resulting from pathogenic organisms.[20] However, with the advent of antibiotics, numerous scientific studies now recognize that sepsis actually results from the dysregulated host immune response to systemic infection, which can also cause organ dysfunction.[21] Based on this knowledge, landmark translational and clinical studies in the 2000s provided strong evidence that early identification of sepsis patients and aggressive infection control and resuscitation were associated with improved mortality (Figure 2, step 1).[22]
Translating Evidence to Care in Sepsis at KPNC
In 2007, the leadership of KPNC initiated a regional effort to improve the quality of care and reduce the variability in performance at its medical centers (Table 1).[23] Reviewing data from nearly 1000 inpatientsthe last 50 consecutive hospital deaths from each of 19 medical centersa mortality diagnostic based on Institute for Healthcare Improvement recommendations[24] revealed that sepsis had a major impact on hospital outcomes. For example, even though sepsis patients were still relatively under‐recognized at the time, accounting for fewer than 3% of hospitalizations, they contributed to one‐quarter of hospital deaths. In light of these compelling data, senior regional leadership identified reducing sepsis mortality as a key performance improvement goal (Figure 2, step 2).
Time Period | Event Summary |
---|---|
| |
2007 | Variability in hospital standardized mortality ratio observed, indicating an opportunity to drive improved outcomes. |
Initiation of staggered implementation of unified electronic medical record across all KP sites (starting in 2006 and ending in 2009). | |
Spring 2008 | Mortality diagnostic chart review completed identifying sepsis and infection‐related causes as key factors in hospital outcomes. |
May 2008 | Regional Mortality Summit held with a focus on patient safety and mortality reduction efforts through performance improvement. Executive regional and local leadership alignment to focus on sepsis performance improvement. |
Summer 2008 | Sepsis Steering Committee evaluates best available evidence, develops treatment algorithms, and plans for medical center pilots. |
Fall 2008 | Pilot intervention deployed at 2 medical centers. |
November 2008 | First Regional Sepsis Summit: development of sepsis performance improvement playbook, training materials, implementation plans, and measurement strategy. |
November 2008 | All medical centers begin to form multidisciplinary sepsis teams and performance improvement committees, obtain equipment and supplies including assembly of a sepsis cart. Multidisciplinary teams included ED physician champion, ED nurse champion, improvement advisor, hospitalists, intensivists, quality improvement personnel, nurse educators, and even resident physicians. |
January 2009 | Performance data collection begins on EGDT processes and outcomes. Initiation of 2 key elements to enhance screening for and detection of sepsis: (1) concomitant ordering of serum lactic acid along with blood cultures, and (2) definition of lactate >2.0 as a critical lab value. |
Use of manual chart review for case finding and central database entry because of ongoing implementation of electronic medical record and limited sepsis‐specific data infrastructure. | |
March 2009 | Regional train the trainer sessions occur and local educational spread efforts begin including: collaborative calls, in‐person training events, and medical center site visits. |
August 2009 | Grant funding from the Gordon and Betty Moore Foundation begins with a planned 2‐year duration providing funding for improvement advisors with performance improvement expertise and data infrastructure development. |
November 2009 | Second Regional Sepsis Summit. Identification of intermediate lactate sepsis patients having significant mortality. |
January 2010 | Initiate measurement of performance for intermediate lactate sepsis patients with a focus on lactate clearance as an outcome measure of interest. |
2010 | Development of an intranet Web‐based data abstraction tool to identify cases and auto‐populate specific fields for review. Facilities were responsible for review of cases at the local level to foster rapid feedback cycles for local performance improvement. Standardized data query tools were deployed to foster local medical center engagement and system‐level evaluation. |
Accompanying development of a sepsis performance improvement scorecard allowing for comparison of longitudinal performance metrics across all facilities. Scorecard elements included: proportion of lactates drawn following ED blood culture, EGDT‐specific bundle elements (ie, number of EGDT cases, antibiotics within 1 hour, first central venous pressure within 2 hours of EGDT start, target mean arterial pressure achievement), repeat lactate elements, balancing measures for central line placement (ie, pneumothorax, central line infection), and overall sepsis statistics. | |
April 2011 | Third Regional Sepsis Summit. Refinement of EGDT bundle and further development of intermediate lactate bundle approach, including piloting specific treatment bundles targeting this population. Collaborative performance improvement environment in which successful strategies at 1 site were rapidly disseminated to other sites including the Sepsis Alert and the Sepsis Clock. |
May 2012 | Research analysis of fluid volume and lactate clearance in intermediate lactate sepsis population begins. |
February 2013 | Fourth Regional Sepsis Summit. Regional spread of intermediate lactate bundle including the use of fluids, antibiotics, and repeat lactate measurements. |
May 2013 | Research analysis of the contribution of sepsis to hospital deaths (within KP and in national sample) as well as post‐sepsis resource utilization and mortality |
March 2014 | Publication of ProCESS randomized clinical trial, requiring systemic reevaluation of EGDT‐based sepsis strategy. Subsequent publications of ARISE and ProMISe trials confirming findings from ProCESS. Updated approach under consideration and informally disseminated to practitioners. |
October 2014 | Updated sepsis treatment guidelines and data capture strategy fully implemented moving away from a catheter‐based strategy for all EGDT‐eligible patients. |
October 2015 | Sixth Regional Sepsis Summit held to adjust sepsis treatment and data measurement strategy to align more closely with CMS SEP‐1 guidelines. |
Based on the principles of performance improvement methodology, clinical and operational leaders established an environment with aligned culture, incentives, and leadership around sepsis care. The effort was launched in late 2008 at a Sepsis Summit, bringing together a multidisciplinary group of stakeholders (eg, hospitalist, emergency department, and intensive care chiefs of staff and nursing managers; medical center and nursing executive and operational leadership) and providing sepsis care pathways based on the best available evidence.[23] Regional investments in the digital infrastructure to support implementation resulted in the provision of granular data within monthly sepsis scorecards quantifying each medical center's performance and trends for a diverse set of sepsis bundle metrics.
The resulting changes in sepsis care were substantial. For example, improved early recognition of infected patients meeting the criteria for sepsis resulted in large changes in the standardized diagnostic criteria used to label patients (Figure 3A). Implementing screening strategies using serum lactate testing for any patient receiving blood cultures resulted in a roughly 10‐fold increase in the use of lactate testing in the emergency department (Figure 3B). Earlier recognition of sepsis also increased the number of patients receiving early antibiotics and receiving central venous catheters for quantitative resuscitation.[23]
CLOSING THE LOOP TOWARD CONTINUOUS LEARNING IN SEPSIS
Leveraging timely and actionable data steeped within an aligned organizational milieu resulted in large‐scale changes across a heterogeneous set of hospitals. However, to realize the true vision of a learning hospital system, a looming question remained: Could the data generated as the byproduct of routine care now be used to complete the virtuous cycle and drive new scientific discovery (Figure 2, step 3)?
Confirming Concordance in the Impact of Sepsis Nationally
The heightened identification of sepsis patients through program implementation revealed that the impact of sepsis on hospital mortality was greater than originally estimated; based on improved patient identification, sepsis now accounted for upward of 1 in 2 hospital deaths.[25] This sobering statistic confirmed that the investments in standardizing best sepsis care following the mortality diagnostic were critical. However, were similar estimates of sepsis‐attributable mortality consistent outside of the KPNC system? To study this, we examined US hospitalizations occurring across >1000 hospitals and involving >6 million hospital stays to estimate corresponding prevalence.[25] In this national sample, sepsis contributed to as many as half of hospital deaths in the United States in 2010, lending strong support to ongoing international and state‐based efforts to improve sepsis care. These studies also paved the way to use these data drawn from our large sepsis population to inform updated international consensus definitions for sepsis and septic shock.[26, 27, 28]
Identifying New Avenues for Reducing the Toll of Sepsis
A major concern of sepsis program leaders was the prior findings that sepsis hospitalizations among Medicare beneficiaries were associated with substantial new cognitive and functional disability.[29] This lingering toll of sepsis had been termed a hidden public health disaster.[30] To further understand the posthospital impact of sepsis and to begin investigating new avenues to reduce this impact, a cohort of patients was followed for 1 year following sepsis hospitalization.[31] Over that period, nearly half of sepsis survivors were rehospitalized. When compared with their presepsis levels of healthcare utilization, middle‐aged and elderly sepsis patients experienced a 3‐fold increase in their days using facility‐based care. Subsequent studies in other populations outside of KPNC have confirmed these concerning findings, resulting in new efforts to address postsepsis survivorship care.[32, 33]
Phenotyping New Targets for Standardized Sepsis Care
At its outset, the sepsis improvement program applied the best available evidence to treat patients with the most severe forms of sepsisseptic shock. However, once the initial implementation phase had succeeded, clinicians and operational leaders quickly realized from the emerging data that there was a far larger group of sepsis patients for whom treatment guidelines were poorly defined.[25, 34, 35] These were severe sepsis patients with so‐called intermediate lactate values between 2 mmol/L and 4 mmol/L; they comprised a substantial proportion of all sepsis patients dying in the hospital. Using data generated from the routine care of sepsis patients treated across 21 hospitals, the sepsis leadership group was able to rapidly assemble a cohort of intermediate lactate sepsis patients up to 20‐ to 100‐fold larger than that reported in prior studies and evaluate their outcomes.[34, 35]
The data used to evaluate these intermediate lactate sepsis patients now spurred a new implementation program in 2013 for a group of patients in whom there was essentially no existing evidence to guide care. Rapidly implemented within a mature sepsis performance improvement program, evaluations at the 6‐month and 1‐year intervals demonstrated significant decreases in mortality.[36] Importantly, to allay the justified concerns of clinicians, these evaluations also clearly showed no evidence of harm from more aggressive fluid resuscitation (eg, increased transfer to intensive care, increased rates of mechanical ventilation). Again, driven by clinician input, subgroup analyses further revealed that the implementation program was only associated with reduced mortality in patients who could be at risk for iatrogenic fluid overload (ie, those with a history of congestive heart failure or chronic kidney disease).[36] Spurred by these provocative findings, operational and clinical leaders are currently considering how to guide future care in these patients, especially with the emerging use of noninvasive methods to quantify patients' fluid responsiveness.
PRINCIPLES FOR LEVERAGING DATA IN THE LEARNING HOSPITAL SYSTEM
The object lesson of using data to drive improved sepsis care and further new scientific discovery offers some important insights for continuous learning.
Building a Digital Infrastructure for Utilizing Granular Hospital Data
As described above, current transitions between the nodes of the learning framework are rife with missed opportunities. Perhaps one of the most glaring is the inability to use highly granular data already collected within the electronic health record (eg, trajectories and trends across vital signs or laboratory results, large‐scale medication administration records to evaluate multidrug interactions). An essential starting point for continuous learning is investing in the digital infrastructure to improve the use of data beyond traditional claims (administrative dataadmission source codes, disposition codes, diagnoses, and procedures). As shown in Table 2, the first key step is incorporating laboratory data into the quality assessment/emmprovement process. In addition, using these data to automate severity of illness and risk adjustment metrics fosters use of similar comparison cohorts across time or disease types.[18, 37, 38, 39, 40]
Data Type | Contents | Degree of Difficulty in Accessing | Degree of Difficulty in Analyzing |
---|---|---|---|
Administrative | Traditional claims data, diagnostic or procedural codes | Low | Low to moderate |
Standard cohort profiling | Limited instances of vitals signs, laboratory, diagnostic testing, or treatment data | Low to moderate | Low to moderate |
Metrics reporting for care improvement | Standard cohort identification, aggregated achievement of treatment targets, scorecard dissemination | Moderate | Moderate |
Advanced cohort profiling | Time series of physiologic data, inpatient triage and treatment data within short temporal intervals | Moderate to high | High |
Research‐grade discovery | Data with breadth (representative sample size) and depth (highly granular physiologic and treatment data) | High | Very high |
Patient‐reported outcomes | Quality of life, functional and cognitive disability | Very high | High |
Employing Novel Methods to Address the Limitations of Using Real‐World Data
The rapid digitization of medicine through the use of electronic medical records offers tremendous opportunities to facilitate continuous learning. However, these opportunities are accompanied by important limitations.[41] Data collected as a byproduct of real‐world care can be vulnerable to many forms of bias and confounding, potentially clouding the validity and robustness of corresponding analytic results. Fortunately, advanced methods including causal inference are now used routinely to address some limitations.[42] In the context of a learning healthcare system, other opportunities for improved study design including cluster randomized trials or stepped wedge implementation can also be employed to preserve the statistical rigor of subsequent analyses.[43] Finally, emerging methods employing randomization through the electronic medical record alongside adaptive trial design offer great potential to increase the efficiency of continuous learning.[44]
Evaluating the Hospital as a Single System
Advances in contemporary hospital care require seamless transitions of patient care, screening strategies, and therapeutic approaches across multiple hospital domains and with diverse providers; these interventions also need to happen rapidly. Many traditional approaches to inpatient care have taken a bottom‐up approach (eg, studying a specific disease within a specific hospital ward like the intensive care unit) that have proven useful but may limit generalizability when applied to a real‐world hospital operating with Pareto optimality (ie, the trade‐off scenario where new resource allocation to 1 area also requires resource withdrawal from another area). In certain cases, an empiric approach, without initial preference for any specific ward or disease, can aid decision making by hospital operational and clinical leaders by providing a global picture of impact and value.
Focusing on Early Detection in Hospital Settings as Secondary Prevention
Once patients have been admitted to the hospital, a race against the clock begins. Each additional hour of hospitalization increases the risks of iatrogenic injury or medical harm manifested by immobility, disorientation and delirium, nosocomial infections, or medication errors, among others. In this context, detection systems that use granular hospital data to focus on the earliest detection of risk can aid critical approaches to secondary prevention (Although the hospitalization for sepsis cannot be avoided, careful attention to mobility can limit the risk of developing delirium. In turn, preventing delirium can limit the risk of new functional disability).
Contextualizing Hospital Care Within a Longitudinal Trajectory
Although we described the benefit of hospital episodes having well‐demarcated beginning and ending points, it remains essential to recognize that the harms associated with hospitalization extend well beyond discharge. In this context, hospitalizations can serve as waypoints in patients' health trajectories as well as an opportunity to achieve patient‐centered care including discussing and aligning goals of care with actual care provision. Furthermore, although we have seen steady declines in hospital mortality over time, it is highly likely that we will reach a nadir in mortality where additional metrics of hospital outcomes will need to include postdischarge events like readmission, long‐term mortality, quality of life, and the prevention of disability or decline.
CONCLUSION
Hospitalizations in the United States are costly and associated with high mortality and morbidity; the toll of hospitalization also extends well beyond hospital discharge. The promise of the learning hospital system has marked improvements in the quality of hospital care, especially where healthcare systems can steep critical investments in data and digital infrastructure within the right culture, incentives, and leadership. Where continuous learning is achieved, data generated during routine care offer the potential to yield new scientific discovery and drive further improvements in hospital care.
Disclosures
As part of our agreement with the Gordon and Betty Moore Foundation, we made a commitment to disseminate our findings in articles such as this one. However, the Foundation and its staff played no role in how we actually structured our articles, nor did they review or preapprove any of the manuscripts submitted as part of the dissemination component. None of the authors has any conflicts of interest to declare of relevance to this work, which was funded by a combination of funding from the Gordon and Betty Moore Foundation, The Permanente Medical Group, Inc., and Kaiser Foundation Hospitals, Inc. VXL was supported by NIH K23GM112018.
In the landmark Best Care at Lower Cost report, the Institute of Medicine presents a compelling vision of a US healthcare system where science, information technology, incentives, and care culture are brought together seamlessly to produce high‐quality healthcare.[1] At the center of this transformation is the learning healthcare system, a system characterized by its ability to leverage data arising from care provision to drive rapid improvements in care delivery.[2] When steeped within the right organizational milieu, these data help to close the virtuous cycle of continuous learning moving from science to evidence to care and back to new science. The anticipated end result is a healthcare system that can provide Americans with superior care at lower cost.
Hospital‐based practitioners will recognize the inpatient setting as an ideal demonstration opportunity for continuous learning. Hospital care is costly, accounting for more than 30% of all US healthcare costs[3]; intensive care alone accounts for a notable proportion of the US gross domestic product.[4] Inpatient care is associated with significant mortality and morbidity, and its use is often greatly increased in patients' last days.[5, 6] Fortunately, the inpatient setting also offers an ideal opportunity to leverage high‐quality data to help inform and improve care. The digitization of medicine means that far more data are now available through electronic health records, medical devices, and tests.[7] This is particularly true for inpatients, for whom a large volume of data are produced even over relatively short hospital stays.
Whereas the challenge to improve hospital care is daunting, there is an incredible opportunity to advance the quality of inpatient care through realizing the vision of the learning hospital system. In the sections that follow, we use an object lessonsepsis care within hospitals of the Kaiser Permanente Northern California (KPNC) integrated healthcare delivery systemto evaluate the challenges and insights gleaned from working toward building a learning hospital system. Then, we describe further steps that could enhance the use of inpatient data to drive improved care.
THE FRAMEWORK OF A LEARNING HEALTHCARE SYSTEM
Best Care at Lower Cost notes a fundamental paradox in US healthcare: although we have witnessed a dramatic expansion in biomedical knowledge, innovative therapies and surgical procedures, and clinical treatments to extend survival, US healthcare persistently falls short on the basic dimensions of quality, outcomes, cost, and equity.[1] The proposed path forward lies in building the learning healthcare system, a system characterized by continuous knowledge development, improvement, and application. Figure 1 shows the critical nodes in the framework for continuous learning, which include: (1) the development of new scientific knowledge (science), (2) the translation of science into clinical evidence of efficacy (evidence), and (3) the application of efficacious interventions through effective care delivery (care). In healthcare today, transitions between these nodes are rife with missed or wasted opportunities like delays in applying high‐quality evidence or poorly managed insights arising from scientific discovery. If such opportunities could be recovered, however, the quality of healthcare could be improved dramatically.[8]
The pursuit of continuous learning is aided by rapid changes in the quality and quantity of biomedical data available over the past decade, especially through the use of electronic health records, novel biomolecular tools, and digital sensors.[2, 7, 9] The Internet has ushered in a new era of data connectivity, for example, allowing for highly engaged communication between patients and providers as well as collaboration between professional or citizen scientists on data of unprecedented scale.[10] New methodologic approaches, including data mining and machine learning, increasingly leverage commodity hardware to conduct previously computationally intractable analyses.[9] Moreover, the development of domain ontologies fosters the discovery of meaningful insights from data of heterogeneous types.[11]
Ultimately, however, improvements in data alone are inadequate to achieve continuous learning. As shown in Figure 1, whereas data form the channels that allow for transitions from science to evidence to care, novel insights need to be steeped within the right culture, motivated by the right incentives, and supported by the right leaders.[1, 12] Within the sustainable learning healthcare system, knowledge generation feeds practice change with the support and guidance of system leadership; improved practice, in turn, generates new knowledge and completes the virtuous cycle of learning.
THE PROMISE OF CONTINUOUS LEARNING IN HOSPITAL SETTINGS
The hospital is an ideal setting in which to foster continuous learning because advances in inpatient care have the potential to substantially improve healthcare quality and value.[8] Americans were hospitalized roughly 37 million times in 2012; in total, these episodes cost $378 billion.[3] Over 700,000 patients die in US hospitals annually, with reports showing that many patients utilize greatly increased inpatient and critical care services near the end of their lives in a manner that appears misaligned with their preferences.[11, 13] Hospital care is also highly variable in quality and cost; this heterogeneity is not closely associated with improved outcomes.[14, 15] Preventable harm and medical injury occur commonly in hospitals and are now recognized to be a leading cause of inpatient death.[16] Finally, emerging research illuminates the substantial toll that acute care has on patients and families resulting in new comorbidity, functional or neuropsychiatric impairment, rehospitalization, and financial burden that persist long after patients are discharged.[17]
Fortunately, inpatient care also exhibits several qualities that improve the likelihood that continuous learning can be achieved. Although it is clear that hospitalizations occur within the arc of a patient's larger health trajectory, these distinct episodes offer the potential to observe patient trajectories and treatments evolving within relatively compressed time intervals; over that same interval, a large volume of data are produced. Stored within comprehensive electronic health records, these granular data now allow inpatient episodes to be digitally recapitulated with high fidelity, bolstering their use in driving care improvements.[18]
AN OBJECT LESSON IN THE LEARNING FRAMEWORK: SEPSIS CARE
Translating Science to Evidence in Sepsis
Although sepsis has attracted great attention in modern hospital care, sepsis was described long ago by Hippocrates to describe the process by which wounds fester.[19] Recast after the confirmation of germ theory, sepsis came to be known primarily as the blood poisoning resulting from pathogenic organisms.[20] However, with the advent of antibiotics, numerous scientific studies now recognize that sepsis actually results from the dysregulated host immune response to systemic infection, which can also cause organ dysfunction.[21] Based on this knowledge, landmark translational and clinical studies in the 2000s provided strong evidence that early identification of sepsis patients and aggressive infection control and resuscitation were associated with improved mortality (Figure 2, step 1).[22]
Translating Evidence to Care in Sepsis at KPNC
In 2007, the leadership of KPNC initiated a regional effort to improve the quality of care and reduce the variability in performance at its medical centers (Table 1).[23] Reviewing data from nearly 1000 inpatientsthe last 50 consecutive hospital deaths from each of 19 medical centersa mortality diagnostic based on Institute for Healthcare Improvement recommendations[24] revealed that sepsis had a major impact on hospital outcomes. For example, even though sepsis patients were still relatively under‐recognized at the time, accounting for fewer than 3% of hospitalizations, they contributed to one‐quarter of hospital deaths. In light of these compelling data, senior regional leadership identified reducing sepsis mortality as a key performance improvement goal (Figure 2, step 2).
Time Period | Event Summary |
---|---|
| |
2007 | Variability in hospital standardized mortality ratio observed, indicating an opportunity to drive improved outcomes. |
Initiation of staggered implementation of unified electronic medical record across all KP sites (starting in 2006 and ending in 2009). | |
Spring 2008 | Mortality diagnostic chart review completed identifying sepsis and infection‐related causes as key factors in hospital outcomes. |
May 2008 | Regional Mortality Summit held with a focus on patient safety and mortality reduction efforts through performance improvement. Executive regional and local leadership alignment to focus on sepsis performance improvement. |
Summer 2008 | Sepsis Steering Committee evaluates best available evidence, develops treatment algorithms, and plans for medical center pilots. |
Fall 2008 | Pilot intervention deployed at 2 medical centers. |
November 2008 | First Regional Sepsis Summit: development of sepsis performance improvement playbook, training materials, implementation plans, and measurement strategy. |
November 2008 | All medical centers begin to form multidisciplinary sepsis teams and performance improvement committees, obtain equipment and supplies including assembly of a sepsis cart. Multidisciplinary teams included ED physician champion, ED nurse champion, improvement advisor, hospitalists, intensivists, quality improvement personnel, nurse educators, and even resident physicians. |
January 2009 | Performance data collection begins on EGDT processes and outcomes. Initiation of 2 key elements to enhance screening for and detection of sepsis: (1) concomitant ordering of serum lactic acid along with blood cultures, and (2) definition of lactate >2.0 as a critical lab value. |
Use of manual chart review for case finding and central database entry because of ongoing implementation of electronic medical record and limited sepsis‐specific data infrastructure. | |
March 2009 | Regional train the trainer sessions occur and local educational spread efforts begin including: collaborative calls, in‐person training events, and medical center site visits. |
August 2009 | Grant funding from the Gordon and Betty Moore Foundation begins with a planned 2‐year duration providing funding for improvement advisors with performance improvement expertise and data infrastructure development. |
November 2009 | Second Regional Sepsis Summit. Identification of intermediate lactate sepsis patients having significant mortality. |
January 2010 | Initiate measurement of performance for intermediate lactate sepsis patients with a focus on lactate clearance as an outcome measure of interest. |
2010 | Development of an intranet Web‐based data abstraction tool to identify cases and auto‐populate specific fields for review. Facilities were responsible for review of cases at the local level to foster rapid feedback cycles for local performance improvement. Standardized data query tools were deployed to foster local medical center engagement and system‐level evaluation. |
Accompanying development of a sepsis performance improvement scorecard allowing for comparison of longitudinal performance metrics across all facilities. Scorecard elements included: proportion of lactates drawn following ED blood culture, EGDT‐specific bundle elements (ie, number of EGDT cases, antibiotics within 1 hour, first central venous pressure within 2 hours of EGDT start, target mean arterial pressure achievement), repeat lactate elements, balancing measures for central line placement (ie, pneumothorax, central line infection), and overall sepsis statistics. | |
April 2011 | Third Regional Sepsis Summit. Refinement of EGDT bundle and further development of intermediate lactate bundle approach, including piloting specific treatment bundles targeting this population. Collaborative performance improvement environment in which successful strategies at 1 site were rapidly disseminated to other sites including the Sepsis Alert and the Sepsis Clock. |
May 2012 | Research analysis of fluid volume and lactate clearance in intermediate lactate sepsis population begins. |
February 2013 | Fourth Regional Sepsis Summit. Regional spread of intermediate lactate bundle including the use of fluids, antibiotics, and repeat lactate measurements. |
May 2013 | Research analysis of the contribution of sepsis to hospital deaths (within KP and in national sample) as well as post‐sepsis resource utilization and mortality |
March 2014 | Publication of ProCESS randomized clinical trial, requiring systemic reevaluation of EGDT‐based sepsis strategy. Subsequent publications of ARISE and ProMISe trials confirming findings from ProCESS. Updated approach under consideration and informally disseminated to practitioners. |
October 2014 | Updated sepsis treatment guidelines and data capture strategy fully implemented moving away from a catheter‐based strategy for all EGDT‐eligible patients. |
October 2015 | Sixth Regional Sepsis Summit held to adjust sepsis treatment and data measurement strategy to align more closely with CMS SEP‐1 guidelines. |
Based on the principles of performance improvement methodology, clinical and operational leaders established an environment with aligned culture, incentives, and leadership around sepsis care. The effort was launched in late 2008 at a Sepsis Summit, bringing together a multidisciplinary group of stakeholders (eg, hospitalist, emergency department, and intensive care chiefs of staff and nursing managers; medical center and nursing executive and operational leadership) and providing sepsis care pathways based on the best available evidence.[23] Regional investments in the digital infrastructure to support implementation resulted in the provision of granular data within monthly sepsis scorecards quantifying each medical center's performance and trends for a diverse set of sepsis bundle metrics.
The resulting changes in sepsis care were substantial. For example, improved early recognition of infected patients meeting the criteria for sepsis resulted in large changes in the standardized diagnostic criteria used to label patients (Figure 3A). Implementing screening strategies using serum lactate testing for any patient receiving blood cultures resulted in a roughly 10‐fold increase in the use of lactate testing in the emergency department (Figure 3B). Earlier recognition of sepsis also increased the number of patients receiving early antibiotics and receiving central venous catheters for quantitative resuscitation.[23]
CLOSING THE LOOP TOWARD CONTINUOUS LEARNING IN SEPSIS
Leveraging timely and actionable data steeped within an aligned organizational milieu resulted in large‐scale changes across a heterogeneous set of hospitals. However, to realize the true vision of a learning hospital system, a looming question remained: Could the data generated as the byproduct of routine care now be used to complete the virtuous cycle and drive new scientific discovery (Figure 2, step 3)?
Confirming Concordance in the Impact of Sepsis Nationally
The heightened identification of sepsis patients through program implementation revealed that the impact of sepsis on hospital mortality was greater than originally estimated; based on improved patient identification, sepsis now accounted for upward of 1 in 2 hospital deaths.[25] This sobering statistic confirmed that the investments in standardizing best sepsis care following the mortality diagnostic were critical. However, were similar estimates of sepsis‐attributable mortality consistent outside of the KPNC system? To study this, we examined US hospitalizations occurring across >1000 hospitals and involving >6 million hospital stays to estimate corresponding prevalence.[25] In this national sample, sepsis contributed to as many as half of hospital deaths in the United States in 2010, lending strong support to ongoing international and state‐based efforts to improve sepsis care. These studies also paved the way to use these data drawn from our large sepsis population to inform updated international consensus definitions for sepsis and septic shock.[26, 27, 28]
Identifying New Avenues for Reducing the Toll of Sepsis
A major concern of sepsis program leaders was the prior findings that sepsis hospitalizations among Medicare beneficiaries were associated with substantial new cognitive and functional disability.[29] This lingering toll of sepsis had been termed a hidden public health disaster.[30] To further understand the posthospital impact of sepsis and to begin investigating new avenues to reduce this impact, a cohort of patients was followed for 1 year following sepsis hospitalization.[31] Over that period, nearly half of sepsis survivors were rehospitalized. When compared with their presepsis levels of healthcare utilization, middle‐aged and elderly sepsis patients experienced a 3‐fold increase in their days using facility‐based care. Subsequent studies in other populations outside of KPNC have confirmed these concerning findings, resulting in new efforts to address postsepsis survivorship care.[32, 33]
Phenotyping New Targets for Standardized Sepsis Care
At its outset, the sepsis improvement program applied the best available evidence to treat patients with the most severe forms of sepsisseptic shock. However, once the initial implementation phase had succeeded, clinicians and operational leaders quickly realized from the emerging data that there was a far larger group of sepsis patients for whom treatment guidelines were poorly defined.[25, 34, 35] These were severe sepsis patients with so‐called intermediate lactate values between 2 mmol/L and 4 mmol/L; they comprised a substantial proportion of all sepsis patients dying in the hospital. Using data generated from the routine care of sepsis patients treated across 21 hospitals, the sepsis leadership group was able to rapidly assemble a cohort of intermediate lactate sepsis patients up to 20‐ to 100‐fold larger than that reported in prior studies and evaluate their outcomes.[34, 35]
The data used to evaluate these intermediate lactate sepsis patients now spurred a new implementation program in 2013 for a group of patients in whom there was essentially no existing evidence to guide care. Rapidly implemented within a mature sepsis performance improvement program, evaluations at the 6‐month and 1‐year intervals demonstrated significant decreases in mortality.[36] Importantly, to allay the justified concerns of clinicians, these evaluations also clearly showed no evidence of harm from more aggressive fluid resuscitation (eg, increased transfer to intensive care, increased rates of mechanical ventilation). Again, driven by clinician input, subgroup analyses further revealed that the implementation program was only associated with reduced mortality in patients who could be at risk for iatrogenic fluid overload (ie, those with a history of congestive heart failure or chronic kidney disease).[36] Spurred by these provocative findings, operational and clinical leaders are currently considering how to guide future care in these patients, especially with the emerging use of noninvasive methods to quantify patients' fluid responsiveness.
PRINCIPLES FOR LEVERAGING DATA IN THE LEARNING HOSPITAL SYSTEM
The object lesson of using data to drive improved sepsis care and further new scientific discovery offers some important insights for continuous learning.
Building a Digital Infrastructure for Utilizing Granular Hospital Data
As described above, current transitions between the nodes of the learning framework are rife with missed opportunities. Perhaps one of the most glaring is the inability to use highly granular data already collected within the electronic health record (eg, trajectories and trends across vital signs or laboratory results, large‐scale medication administration records to evaluate multidrug interactions). An essential starting point for continuous learning is investing in the digital infrastructure to improve the use of data beyond traditional claims (administrative dataadmission source codes, disposition codes, diagnoses, and procedures). As shown in Table 2, the first key step is incorporating laboratory data into the quality assessment/emmprovement process. In addition, using these data to automate severity of illness and risk adjustment metrics fosters use of similar comparison cohorts across time or disease types.[18, 37, 38, 39, 40]
Data Type | Contents | Degree of Difficulty in Accessing | Degree of Difficulty in Analyzing |
---|---|---|---|
Administrative | Traditional claims data, diagnostic or procedural codes | Low | Low to moderate |
Standard cohort profiling | Limited instances of vitals signs, laboratory, diagnostic testing, or treatment data | Low to moderate | Low to moderate |
Metrics reporting for care improvement | Standard cohort identification, aggregated achievement of treatment targets, scorecard dissemination | Moderate | Moderate |
Advanced cohort profiling | Time series of physiologic data, inpatient triage and treatment data within short temporal intervals | Moderate to high | High |
Research‐grade discovery | Data with breadth (representative sample size) and depth (highly granular physiologic and treatment data) | High | Very high |
Patient‐reported outcomes | Quality of life, functional and cognitive disability | Very high | High |
Employing Novel Methods to Address the Limitations of Using Real‐World Data
The rapid digitization of medicine through the use of electronic medical records offers tremendous opportunities to facilitate continuous learning. However, these opportunities are accompanied by important limitations.[41] Data collected as a byproduct of real‐world care can be vulnerable to many forms of bias and confounding, potentially clouding the validity and robustness of corresponding analytic results. Fortunately, advanced methods including causal inference are now used routinely to address some limitations.[42] In the context of a learning healthcare system, other opportunities for improved study design including cluster randomized trials or stepped wedge implementation can also be employed to preserve the statistical rigor of subsequent analyses.[43] Finally, emerging methods employing randomization through the electronic medical record alongside adaptive trial design offer great potential to increase the efficiency of continuous learning.[44]
Evaluating the Hospital as a Single System
Advances in contemporary hospital care require seamless transitions of patient care, screening strategies, and therapeutic approaches across multiple hospital domains and with diverse providers; these interventions also need to happen rapidly. Many traditional approaches to inpatient care have taken a bottom‐up approach (eg, studying a specific disease within a specific hospital ward like the intensive care unit) that have proven useful but may limit generalizability when applied to a real‐world hospital operating with Pareto optimality (ie, the trade‐off scenario where new resource allocation to 1 area also requires resource withdrawal from another area). In certain cases, an empiric approach, without initial preference for any specific ward or disease, can aid decision making by hospital operational and clinical leaders by providing a global picture of impact and value.
Focusing on Early Detection in Hospital Settings as Secondary Prevention
Once patients have been admitted to the hospital, a race against the clock begins. Each additional hour of hospitalization increases the risks of iatrogenic injury or medical harm manifested by immobility, disorientation and delirium, nosocomial infections, or medication errors, among others. In this context, detection systems that use granular hospital data to focus on the earliest detection of risk can aid critical approaches to secondary prevention (Although the hospitalization for sepsis cannot be avoided, careful attention to mobility can limit the risk of developing delirium. In turn, preventing delirium can limit the risk of new functional disability).
Contextualizing Hospital Care Within a Longitudinal Trajectory
Although we described the benefit of hospital episodes having well‐demarcated beginning and ending points, it remains essential to recognize that the harms associated with hospitalization extend well beyond discharge. In this context, hospitalizations can serve as waypoints in patients' health trajectories as well as an opportunity to achieve patient‐centered care including discussing and aligning goals of care with actual care provision. Furthermore, although we have seen steady declines in hospital mortality over time, it is highly likely that we will reach a nadir in mortality where additional metrics of hospital outcomes will need to include postdischarge events like readmission, long‐term mortality, quality of life, and the prevention of disability or decline.
CONCLUSION
Hospitalizations in the United States are costly and associated with high mortality and morbidity; the toll of hospitalization also extends well beyond hospital discharge. The promise of the learning hospital system has marked improvements in the quality of hospital care, especially where healthcare systems can steep critical investments in data and digital infrastructure within the right culture, incentives, and leadership. Where continuous learning is achieved, data generated during routine care offer the potential to yield new scientific discovery and drive further improvements in hospital care.
Disclosures
As part of our agreement with the Gordon and Betty Moore Foundation, we made a commitment to disseminate our findings in articles such as this one. However, the Foundation and its staff played no role in how we actually structured our articles, nor did they review or preapprove any of the manuscripts submitted as part of the dissemination component. None of the authors has any conflicts of interest to declare of relevance to this work, which was funded by a combination of funding from the Gordon and Betty Moore Foundation, The Permanente Medical Group, Inc., and Kaiser Foundation Hospitals, Inc. VXL was supported by NIH K23GM112018.
- Institute of Medicine. Best Care at Lower Cost: The Path to Continuously Learning Health Care in America. Washington, DC: The National Academies Press; 2012.
- Toward a science of learning systems: a research agenda for the high‐functioning Learning Health System. J Am Med Inform Assoc. 2015;22(1):43–50. , , , et al.
- National Center for Health Statistics. Health, United States, 2014: With Special Feature on Adults Aged 55–64. Hyattsville, MD; 2015.
- Critical care medicine in the United States 2000‐2005: an analysis of bed numbers, occupancy rates, payer mix, and costs. Crit Care Med. 2010;38(1):65–71. , .
- Trends and variation in end‐of‐life care for medicare beneficiaries with severe chronic illness. A report of the Dartmouth Atlas Project. Lebanon, NH: The Dartmouth Institute for Health Policy and Clinical Practice; 2011. , , , .
- Change in end‐of‐life care for Medicare beneficiaries: site of death, place of care, and health care transitions in 2000, 2005, and 2009. JAMA. 2013;309(5):470–477. , , , et al.
- Finding the missing link for big biomedical data. JAMA. 2014;311(24):2479–2480. , , .
- Code red and blue—safely limiting health care's GDP footprint. N Engl J Med. 2013;368(1):1–3. .
- The inevitable application of big data to health care. JAMA. 2013;309(13):1351–1352. , .
- What is citizen science?—a scientometric meta‐analysis. PLoS One. 2016;11(1):e0147152. , .
- Biomedical ontologies: a functional perspective. Brief Bioinform. 2008;9(1):75–90. , , .
- Rapid learning: a breakthrough agenda. Health Aff (Millwood). 2014;33(7):1155–1162. .
- Are regional variations in end‐of‐life care intensity explained by patient preferences?: a study of the US Medicare population. Med Care. 2007;45(5):386–393. , , , et al.
- Extreme markup: the fifty US hospitals with the highest charge‐to‐cost ratios. Health Aff (Millwood). 2015;34(6):922–928. , .
- The price ain't right? Hospital prices and health spending on the privately insured. Health Care Pricing Project website. Available at: http://www.healthcarepricingproject.org/sites/default/files/pricing_variation_manuscript_0.pdf. Accessed February 15, 2016 , , , .
- A new, evidence‐based estimate of patient harms associated with hospital care. J Patient Saf. 2013;9(3):122–128. .
- Hospitalization‐associated disability: “she was probably able to ambulate, but I'm not sure”. JAMA. 2011;306(16):1782–1793. , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated health care delivery system. Med Care. 2013;51(5):446–453. , , , , .
- New definitions for sepsis and septic shock: continuing evolution but with much still to be done. JAMA. 2016;315(8):757–759. .
- Severe sepsis and septic shock. N Engl J Med. 2013;369(21):2063. , .
- Sepsis‐induced immunosuppression: from cellular dysfunctions to immunotherapy. Nat Rev Immunol. 2013;13(12):862–874. , , .
- Early goal‐directed therapy in the treatment of severe sepsis and septic shock. N Engl J Med. 2001;345(19):1368–1377. , , , et al.
- Kaiser Permanente's performance improvement system, part 3: multisite improvements in care for patients with sepsis. Jt Comm J Qual Patient Saf. 2011;37(11):483–493. , , , et al.
- Understanding the components of quality improvement collaboratives: a systematic literature review. Milbank Q. 2013;91(2):354–394. , , , , .
- Hospital deaths in patients with sepsis from 2 independent cohorts. JAMA. 2014;312(1):90–92. , , , et al.
- Assessment of clinical criteria for sepsis: for the Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):762–774. , , , et al.
- Developing a new definition and assessing new clinical criteria for septic shock: for the Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):775–787. , , , et al.
- The Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):801–810. , , , et al.
- Long‐term cognitive impairment and functional disability among survivors of severe sepsis. JAMA. 2010;304(16):1787–1794. , , , .
- The lingering consequences of sepsis: a hidden public health disaster? JAMA. 2010;304(16):1833–1834. .
- Hospital readmission and healthcare utilization following sepsis in community settings. J Hosp Med. 2014;9(8):502–507. , , , , , .
- Increased 1‐year healthcare use in survivors of severe sepsis. Am J Respir Crit Care Med. 2014;190(1):62–69. , , , , .
- Post‐acute care use and hospital readmission after sepsis. Ann Am Thorac Soc. 2015;12(6):904–913. , , , et al.
- Fluid volume, lactate values, and mortality in sepsis patients with intermediate lactate values. Ann Am Thorac Soc. 2013;10(5):466–473. , , , , .
- Prognosis of emergency department patients with suspected infection and intermediate lactate levels: a systematic review. J Crit Care. 2014;29(3):334–339. , , .
- Multicenter implementation of a treatment bundle for sepsis patients with intermediate lactate values. Am J Respir Crit Care Med. 2016;193(11):1264–1270. , , , et al.
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases. Am J Manag Care. 2008;14(3):158–166. , , , et al.
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6(2):74–80. , , , , , .
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- An electronic Simplified Acute Physiology Score‐based risk adjustment score for critical illness in an integrated healthcare system. Crit Care Med. 2013;41(1):41–48. , , , , .
- Learning from big health care data. N Engl J Med. 2014;370(23):2161–2163. .
- Getting the methods right—the foundation of patient‐centered outcomes research. N Engl J Med. 2012;367(9):787–790. , .
- The stepped wedge cluster randomised trial: rationale, design, analysis, and reporting. BMJ. 2015;350:h391. , , , , .
- Fusing randomized trials with big data: the key to self‐learning health care systems? JAMA. 2015;314(8):767–768. .
- Institute of Medicine. Best Care at Lower Cost: The Path to Continuously Learning Health Care in America. Washington, DC: The National Academies Press; 2012.
- Toward a science of learning systems: a research agenda for the high‐functioning Learning Health System. J Am Med Inform Assoc. 2015;22(1):43–50. , , , et al.
- National Center for Health Statistics. Health, United States, 2014: With Special Feature on Adults Aged 55–64. Hyattsville, MD; 2015.
- Critical care medicine in the United States 2000‐2005: an analysis of bed numbers, occupancy rates, payer mix, and costs. Crit Care Med. 2010;38(1):65–71. , .
- Trends and variation in end‐of‐life care for medicare beneficiaries with severe chronic illness. A report of the Dartmouth Atlas Project. Lebanon, NH: The Dartmouth Institute for Health Policy and Clinical Practice; 2011. , , , .
- Change in end‐of‐life care for Medicare beneficiaries: site of death, place of care, and health care transitions in 2000, 2005, and 2009. JAMA. 2013;309(5):470–477. , , , et al.
- Finding the missing link for big biomedical data. JAMA. 2014;311(24):2479–2480. , , .
- Code red and blue—safely limiting health care's GDP footprint. N Engl J Med. 2013;368(1):1–3. .
- The inevitable application of big data to health care. JAMA. 2013;309(13):1351–1352. , .
- What is citizen science?—a scientometric meta‐analysis. PLoS One. 2016;11(1):e0147152. , .
- Biomedical ontologies: a functional perspective. Brief Bioinform. 2008;9(1):75–90. , , .
- Rapid learning: a breakthrough agenda. Health Aff (Millwood). 2014;33(7):1155–1162. .
- Are regional variations in end‐of‐life care intensity explained by patient preferences?: a study of the US Medicare population. Med Care. 2007;45(5):386–393. , , , et al.
- Extreme markup: the fifty US hospitals with the highest charge‐to‐cost ratios. Health Aff (Millwood). 2015;34(6):922–928. , .
- The price ain't right? Hospital prices and health spending on the privately insured. Health Care Pricing Project website. Available at: http://www.healthcarepricingproject.org/sites/default/files/pricing_variation_manuscript_0.pdf. Accessed February 15, 2016 , , , .
- A new, evidence‐based estimate of patient harms associated with hospital care. J Patient Saf. 2013;9(3):122–128. .
- Hospitalization‐associated disability: “she was probably able to ambulate, but I'm not sure”. JAMA. 2011;306(16):1782–1793. , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated health care delivery system. Med Care. 2013;51(5):446–453. , , , , .
- New definitions for sepsis and septic shock: continuing evolution but with much still to be done. JAMA. 2016;315(8):757–759. .
- Severe sepsis and septic shock. N Engl J Med. 2013;369(21):2063. , .
- Sepsis‐induced immunosuppression: from cellular dysfunctions to immunotherapy. Nat Rev Immunol. 2013;13(12):862–874. , , .
- Early goal‐directed therapy in the treatment of severe sepsis and septic shock. N Engl J Med. 2001;345(19):1368–1377. , , , et al.
- Kaiser Permanente's performance improvement system, part 3: multisite improvements in care for patients with sepsis. Jt Comm J Qual Patient Saf. 2011;37(11):483–493. , , , et al.
- Understanding the components of quality improvement collaboratives: a systematic literature review. Milbank Q. 2013;91(2):354–394. , , , , .
- Hospital deaths in patients with sepsis from 2 independent cohorts. JAMA. 2014;312(1):90–92. , , , et al.
- Assessment of clinical criteria for sepsis: for the Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):762–774. , , , et al.
- Developing a new definition and assessing new clinical criteria for septic shock: for the Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):775–787. , , , et al.
- The Third International Consensus Definitions for Sepsis and Septic Shock (Sepsis‐3). JAMA. 2016;315(8):801–810. , , , et al.
- Long‐term cognitive impairment and functional disability among survivors of severe sepsis. JAMA. 2010;304(16):1787–1794. , , , .
- The lingering consequences of sepsis: a hidden public health disaster? JAMA. 2010;304(16):1833–1834. .
- Hospital readmission and healthcare utilization following sepsis in community settings. J Hosp Med. 2014;9(8):502–507. , , , , , .
- Increased 1‐year healthcare use in survivors of severe sepsis. Am J Respir Crit Care Med. 2014;190(1):62–69. , , , , .
- Post‐acute care use and hospital readmission after sepsis. Ann Am Thorac Soc. 2015;12(6):904–913. , , , et al.
- Fluid volume, lactate values, and mortality in sepsis patients with intermediate lactate values. Ann Am Thorac Soc. 2013;10(5):466–473. , , , , .
- Prognosis of emergency department patients with suspected infection and intermediate lactate levels: a systematic review. J Crit Care. 2014;29(3):334–339. , , .
- Multicenter implementation of a treatment bundle for sepsis patients with intermediate lactate values. Am J Respir Crit Care Med. 2016;193(11):1264–1270. , , , et al.
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases. Am J Manag Care. 2008;14(3):158–166. , , , et al.
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6(2):74–80. , , , , , .
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- An electronic Simplified Acute Physiology Score‐based risk adjustment score for critical illness in an integrated healthcare system. Crit Care Med. 2013;41(1):41–48. , , , , .
- Learning from big health care data. N Engl J Med. 2014;370(23):2161–2163. .
- Getting the methods right—the foundation of patient‐centered outcomes research. N Engl J Med. 2012;367(9):787–790. , .
- The stepped wedge cluster randomised trial: rationale, design, analysis, and reporting. BMJ. 2015;350:h391. , , , , .
- Fusing randomized trials with big data: the key to self‐learning health care systems? JAMA. 2015;314(8):767–768. .
Healthcare Utilization after Sepsis
Sepsis, the systemic inflammatory response to infection, is a major public health concern.[1] Worldwide, sepsis affects millions of hospitalized patients each year.[2] In the United States, it is the single most expensive cause of hospitalization.[3, 4, 5, 6] Multiple studies suggest that sepsis hospitalizations are also increasing in frequency.[3, 6, 7, 8, 9, 10]
Improved sepsis care has dramatically reduced in‐hospital mortality.[11, 12, 13] However, the result is a growing number of sepsis survivors discharged with new disability.[1, 9, 14, 15, 16] Despite being a common cause of hospitalization, little is known about how to improve postsepsis care.[15, 17, 18, 19] This contrasts with other, often less common, hospital conditions for which many studies evaluating readmission and postdischarge care are available.[20, 21, 22, 23] Identifying the factors contributing to high utilization could lend critical insight to designing interventions that improve long‐term sepsis outcomes.[24]
We conducted a retrospective study of sepsis patients discharged in 2010 at Kaiser Permanente Northern California (KPNC) to describe their posthospital trajectories. In this diverse community‐hospitalbased population, we sought to identify the patient‐level factors that impact the posthospital healthcare utilization of sepsis survivors.
METHODS
This study was approved by the KPNC institutional review board.
Setting
We conducted a retrospective study of sepsis patients aged 18 years admitted to KPNC hospitals in 2010 whose hospitalizations included an overnight stay, began in a KPNC hospital, and was not for peripartum care. We identified sepsis based on International Classification of Disease, 9th Edition principal diagnosis codes used at KPNC, which capture a similar population to that from the Angus definition (see Supporting Appendix, Table 1, in the online version of this article).[7, 25, 26] We denoted each patient's first sepsis hospitalization as the index event.
Predicted Hospital Mortality Quartiles (n=1,586 for Each Group) | |||||
---|---|---|---|---|---|
Overall | 1 | 2 | 3 | 4 | |
| |||||
Baseline | |||||
Age, y, mean | 71.915.7 | 62.317.8 | 71.214.2 | 75.612.7 | 78.612.2 |
By age category | |||||
<45 years | 410 (6.5) | 290 (18.3) | 71 (4.5) | 25 (1.6) | 24 (1.5) |
4564 years | 1,425 (22.5) | 539 (34.0) | 407 (25.7) | 292 (18.4) | 187 (11.8) |
6584 years | 3,036 (47.9) | 601 (37.9) | 814 (51.3) | 832 (52.5) | 789 (49.8) |
85 years | 1,473 (23.2) | 156 (9.8) | 294 (18.5) | 437 (27.6) | 586 (37.0) |
Male | 2,973 (46.9) | 686 (43.3) | 792 (49.9) | 750 (47.3) | 745 (47.0) |
Comorbidity | |||||
COPS2 score | 5143 | 2627 | 5441 | 6445 | 6245 |
Charlson score | 2.01.5 | 1.31.2 | 2.11.4 | 2.41.5 | 2.41.5 |
Hospitalization | |||||
LAPS2 severity score | 10742 | 6621 | 9020 | 11423 | 15928 |
Admitted via emergency department | 6,176 (97.4) | 1,522 (96.0) | 1,537 (96.9) | 1,539 (97.0) | 1,578 (99.5) |
Direct ICU admission | 1,730 (27.3) | 169 (10.7) | 309 (19.5) | 482 (30.4) | 770 (48.6) |
ICU transfer, at any time | 2,206 (34.8) | 279 (17.6) | 474 (29.9) | 603 (38.0) | 850 (53.6) |
Hospital mortality | |||||
Predicted, % | 10.513.8 | 1.00.1 | 3.40.1 | 8.32.3 | 29.415.8 |
Observed | 865 (13.6) | 26 (1.6) | 86 (5.4) | 197 (12.4) | 556 (35.1) |
Hospital length of stay, d | 5.86.4 | 4.43.8 | 5.45.7 | 6.68.0 | 6.66.9 |
We linked hospital episodes with existing KPNC inpatient databases to describe patient characteristics.[27, 28, 29, 30] We categorized patients by age (45, 4564, 6584, and 85 years) and used Charlson comorbidity scores and Comorbidity Point Scores 2 (COPS2) to quantify comorbid illness burden.[28, 30, 31, 32] We quantified acute severity of illness using the Laboratory Acute Physiology Scores 2 (LAPS2), which incorporates 15 laboratory values, 5 vital signs, and mental status prior to hospital admission (including emergency department data).[30] Both the COPS2 and LAPS2 are independently associated with hospital mortality.[30, 31] We also generated a summary predicted risk of hospital mortality based on a validated risk model and stratified patients by quartiles.[30] We determined whether patients were admitted to the intensive care unit (ICU).[29]
Outcomes
We used patients' health insurance administrative data to quantify postsepsis utilization. Within the KPNC integrated healthcare delivery system, uniform information systems capture all healthcare utilization of insured members including services received at non‐KPNC facilities.[28, 30] We collected utilization data from the year preceding index hospitalization (presepsis) and for the year after discharge date or until death (postsepsis). We ascertained mortality after discharge from KPNC medical records as well as state and national death record files.
We grouped services into facility‐based or outpatient categories. Facility‐based services included inpatient admission, subacute nursing facility or long‐term acute care, and emergency department visits. We grouped outpatient services as hospice, home health, outpatient surgery, clinic, or other (eg, laboratory). We excluded patients whose utilization records were not available over the full presepsis interval. Among these 1211 patients (12.5% of total), the median length of records prior to index hospitalization was 67 days, with a mean value of 117 days.
Statistical Analysis
Our primary outcomes of interest were hospital readmission and utilization in the year after sepsis. We defined a hospital readmission as any inpatient stay after the index hospitalization grouped within 1‐, 3‐, 6‐, and 12‐month intervals. We designated those within 30 days as an early readmission. We grouped readmission principal diagnoses, where available, by the 17 Healthcare Cost and Utilization Project (HCUP) Clinical Classifications Software multilevel categories with sepsis in the infectious category.[33, 34] In secondary analysis, we also designated other infectious diagnoses not included in the standard HCUP infection category (eg, pneumonia, meningitis, cellulitis) as infection (see Supporting Appendix in the online version of this article).
We quantified outpatient utilization based on the number of episodes recorded. For facility‐based utilization, we calculated patient length of stay intervals. Because patients surviving their index hospitalization might not survive the entire year after discharge, we also calculated utilization adjusted for patients' living days by dividing the total facility length of stay by the number of living days after discharge.
Continuous data are represented as mean (standard deviation [SD]) and categorical data as number (%). We compared groups with analysis of variance or 2 testing. We estimated survival with Kaplan‐Meier analysis (95% confidence interval) and compared groups with log‐rank testing. We compared pre‐ and postsepsis healthcare utilization with paired t tests.
To identify factors associated with early readmission after sepsis, we used a competing risks regression model.[35] The dependent variable was time to readmission and the competing hazard was death within 30 days without early readmission; patients without early readmission or death were censored at 30 days. The independent variables included age, gender, comorbid disease burden (COPS2), acute severity of illness (LAPS2), any use of intensive care, total index length of stay, and percentage of living days prior to sepsis hospitalization spent utilizing facility‐based care. We also used logistic regression to quantify the association between these variables and high postsepsis utilization; we defined high utilization as 15% of living days postsepsis spent in facility‐based care. For each model, we quantified the relative contribution of each predictor variable to model performance based on differences in log likelihoods.[35, 36] We conducted analyses using STATA/SE version 11.2 (StataCorp, College Station, TX) and considered a P value of <0.05 to be significant.
RESULTS
Cohort Characteristics
Our study cohort included 6344 patients with index sepsis hospitalizations in 2010 (Table 1). Mean age was 72 (SD 16) years including 1835 (28.9%) patients aged <65 years. During index hospitalizations, higher predicted mortality was associated with increased age, comorbid disease burden, and severity of illness (P<0.01 for each). ICU utilization increased across predicted mortality strata; for example, 10.7% of patients in the lowest quartile were admitted directly to the ICU compared with 48.6% in the highest quartile. In the highest quartile, observed mortality was 35.1%.
One‐Year Survival
A total of 5479 (86.4%) patients survived their index sepsis hospitalization. Overall survival after living discharge was 90.5% (range, 89.6%91.2%) at 30 days and 71.3% (range, 70.1%72.5%) at 1 year. However, postsepsis survival was strongly modified by age (Figure 1). For example, 1‐year survival was 94.1% (range, 91.2%96.0%) for <45 year olds and 54.4% (range, 51.5%57.2%) for 85 year olds (P<0.01). Survival was also modified by predicted mortality, however, not by ICU admission during index hospitalization (P=0.18) (see Supporting Appendix, Figure 1, in the online version of this article).
Hospital Readmission
Overall, 978 (17.9%) patients had early readmission after index discharge (Table 2); nearly half were readmitted at least once in the year following discharge. Rehospitalization frequency was slightly lower when including patients with incomplete presepsis data (see Supporting Appendix, Table 2, in the online version of this article). The frequency of hospital readmission varied based on patient age and severity of illness. For example, 22.3% of patients in the highest predicted mortality quartile had early readmission compared with 11.6% in the lowest. The median time from discharge to early readmission was 11 days. Principal diagnoses were available for 78.6% of all readmissions (see Supporting Appendix, Table 3, in the online version of this article). Between 28.3% and 42.7% of those readmissions were for infectious diagnoses (including sepsis).
Predicted Mortality Quartile | |||||
---|---|---|---|---|---|
Readmission | Overall | 1 | 2 | 3 | 4 |
Within 30 days | 978 (17.9) | 158 (11.6) | 242 (17.7) | 274 (20.0) | 304 (22.3) |
Within 90 days | 1,643 (30.1) | 276 (20.2) | 421 (30.8) | 463 (33.9) | 483 (35.4) |
Within 180 days | 2,061 (37.7) | 368 (26.9) | 540 (39.5) | 584 (42.7) | 569 (41.7) |
Within 365 days | 2,618 (47.9) | 498 (36.4) | 712 (52.1) | 723 (52.9) | 685 (50.2) |
Variable | Hazard Ratio for Early Readmission | Odds Ratio for High Utilization | ||
---|---|---|---|---|
HR (95% CI) | Relative Contribution | OR (95% CI) | Relative Contribution | |
| ||||
Age category | 1.2% | 11.1% | ||
<45 years | 1.00 [reference] | 1.00 [reference] | ||
4564 years | 0.86 (0.64‐1.16) | 2.22 (1.30‐3.83)a | ||
6584 years | 0.92 (0.69‐1.21) | 3.66 (2.17‐6.18)a | ||
85 years | 0.95 (0.70‐1.28) | 4.98 (2.92‐8.50)a | ||
Male | 0.99 (0.88‐1.13) | 0.0% | 0.86 (0.74‐1.00) | 0.1% |
Severity of illness (LAPS2) | 1.08 (1.04‐1.12)a | 12.4% | 1.22 (1.17‐1.27)a | 11.3% |
Comorbid illness (COPS2) | 1.16 (1.12‐1.19)a | 73.9% | 1.13 (1.09‐1.17)a | 5.9% |
Intensive care | 1.21 (1.05‐1.40)a | 5.2% | 1.02 (0.85‐1.21) | 0.0% |
Hospital length of stay, day | 1.01 (1.001.02)b | 6.6% | 1.04 (1.03‐1.06)a | 6.9% |
Prior utilization, per 10% | 0.98 (0.95‐1.02) | 0.7% | 1.74 (1.61‐1.88)a | 64.2% |
Healthcare Utilization
The unadjusted difference between pre‐ and postsepsis healthcare utilization among survivors was statistically significant for most categories but of modest clinical significance (see Supporting Appendix, Table 4, in the online version of this article). For example, the mean number of presepsis hospitalizations was 0.9 (1.4) compared to 1.0 (1.5) postsepsis (P<0.01). After adjusting for postsepsis living days, the difference in utilization was more pronounced (Figure 2). Overall, there was roughly a 3‐fold increase in the mean percentage of living days spent in facility‐based care between patients' pre‐ and postsepsis phases (5.3% vs 15.0%, P<0.01). Again, the difference was strongly modified by age. For patients aged <45 years, the difference was not statistically significant (2.4% vs 2.9%, P=0.32), whereas for those aged 65 years, it was highly significant (6.2% vs 18.5%, P<0.01).
Factors associated with early readmission included severity of illness, comorbid disease burden, index hospital length of stay, and intensive care (Table 3). However, the dominant factor explaining variation in the risk of early readmission was patients' prior comorbid disease burden (73.9%), followed by acute severity of illness (12.4%), total hospital length of stay (6.6%), and the need for intensive care (5.2%). Severity of illness and age were also significantly associated with higher odds of high postsepsis utilization; however, the dominant factor contributing to this risk was a history of high presepsis utilization (64.2%).
DISCUSSION
In this population‐based study in a community healthcare system, the impact of sepsis extended well beyond the initial hospitalization. One in 6 sepsis survivors was readmitted within 30 days, and roughly half were readmitted within 1 year. Fewer than half of rehospitalizations were for sepsis. Patients had a 3‐fold increase in the percentage of living days spent in hospitals or care facilities after sepsis hospitalization. Although age and acute severity of illness strongly modified healthcare utilization and mortality after sepsis, the dominant factors contributing to early readmission and high utilization ratescomorbid disease burden and presepsis healthcare utilizationwere present prior to hospitalization.
Sepsis is the single most expensive cause of US hospitalizations.[3, 4, 5] Despite its prevalence, there are little contemporary data identifying factors that impact healthcare utilization among sepsis survivors.[9, 16, 17, 19, 24, 36, 37] Recently, Prescott and others found that in Medicare beneficiaries, following severe sepsis, healthcare utilization was markedly increased.[17] More than one‐quarter of survivors were readmitted within 30 days, and 63.8% were readmitted within a year. Severe sepsis survivors also spent an average of 26% of their living days in a healthcare facility, a nearly 4‐fold increase compared to their presepsis phase. The current study included a population with a broader age and severity range; however, in a similar subgroup of patients, for those aged 65 years within the highest predicted mortality quartile, the frequency of readmission was similar. These findings are concordant with those from prior studies.[17, 19, 36, 37]
Among sepsis survivors, most readmissions were not for sepsis or infectious diagnoses, which is a novel finding with implications for designing approaches to reduce rehospitalization. The pattern in sepsis is similar to that seen in other common and costly hospital conditions.[17, 20, 23, 38, 39, 40] For example, between 18% and 25% of Medicare beneficiaries hospitalized for heart failure, acute myocardial infarction, or pneumonia were readmitted within 30 days; fewer than one‐third had the same diagnosis.[20] The timing of readmission in our sepsis cohort was also similar to that seen in other conditions.[20] For example, the median time of early readmission in this study was 11 days; it was between 10 and 12 days for patients with heart failure, pneumonia, and myocardial infarction.[20]
Krumholz and others suggest that the pattern of early rehospitalization after common acute conditions reflects a posthospital syndromean acquired, transient period of vulnerabilitythat could be the byproduct of common hospital factors.[20, 41] Such universal impairments might result from new physical and neurocognitive disability, nutritional deficiency, and sleep deprivation or delirium, among others.[41] If this construct were also true in sepsis, it could have important implications on the design of postsepsis care. However, prior studies suggest that sepsis patients may be particularly vulnerable to the sequelae of hospitalization.[2, 42, 43, 44, 45]
Among Medicare beneficiaries, Iwashyna and others reported that hospitalizations for severe sepsis resulted in significant increases in physical limitations and moderate to severe cognitive impairment.[1, 14, 46] Encephalopathy, sleep deprivation, and delirium are also frequently seen in sepsis patients.[47, 48] Furthermore, sepsis patients frequently need intensive care, which is also associated with increased patient disability and injury.[16, 46, 49, 50] We found that severity of illness and the need for intensive care were both predictive of the need for early readmission following sepsis. We also confirmed the results of prior studies suggesting that sepsis outcomes are strongly modified by age.[16, 19, 43, 51]
However, we found that the dominant factors contributing to patients' health trajectories were conditions present prior to admission. This finding is in accord with prior suggestions that acute severity of illness only partially predicts patients facing adverse posthospital sequelae.[23, 41, 52] Among sepsis patients, prior work demonstrates that inadequate consideration for presepsis level of function and utilization can result in an overestimation of the impact of sepsis on postdischarge health.[52, 53] Further, we found that the need for intensive care was not independently associated with an increased risk of high postsepsis utilization after adjusting for illness severity, a finding also seen in prior studies.[17, 23, 38, 51]
Taken together, our findings might suggest that an optimal approach to posthospital care in sepsis should focus on treatment approaches that address disease‐specific problems within the much larger context of common hospital risks. However, further study is necessary to clearly define the mechanisms by which age, severity of illness, and intensive care affect subsequent healthcare utilization. Furthermore, sepsis patients are a heterogeneous population in terms of severity of illness, site and pathogen of infection, and underlying comorbidity whose posthospital course remains incompletely characterized, limiting our ability to draw strong inferences.
These results should be interpreted in light of the study's limitations. First, our cohort included patients with healthcare insurance within a community‐based healthcare system. Care within the KPNC system, which bears similarities with accountable care organizations, is enhanced through service integration and a comprehensive health information system. Although prior studies suggest that these characteristics result in improved population‐based care, it is unclear whether there is a similar impact in hospital‐based conditions such as sepsis.[54, 55] Furthermore, care within an integrated system may impact posthospital utilization patterns and could limit generalizability. However, prior studies demonstrate the similarity of KPNC members to other patients in the same region in terms of age, socioeconomics, overall health behaviors, and racial/ethnic diversity.[56] Second, our study did not characterize organ dysfunction based on diagnosis coding, a common feature of sepsis studies that lack detailed physiologic severity data.[4, 5, 6, 8, 26] Instead, we focused on using granular laboratory and vital signs data to ensure accurate risk adjustment using a validated system developed in >400,000 hospitalizations.[30] Although this method may hamper comparisons with existing studies, traditional methods of grading severity by diagnosis codes can be vulnerable to biases resulting in wide variability.[10, 23, 26, 57, 58] Nonetheless, it is likely that characterizing preexisting and acute organ dysfunction will improve risk stratification in the heterogeneous sepsis population. Third, this study did not include data regarding patients' functional status, which has been shown to strongly predict patient outcomes following hospitalization. Fourth, this study did not address the cost of care following sepsis hospitalizations.[19, 59] Finally, our study excluded patients with incomplete utilization records, a choice designed to avoid the spurious inferences that can result from such comparisons.[53]
In summary, we found that sepsis exacted a considerable toll on patients in the hospital and in the year following discharge. Sepsis patients were frequently rehospitalized within a month of discharge, and on average had a 3‐fold increase in their subsequent time spent in healthcare facilities. Although age, severity of illness, and the need for ICU care impacted postsepsis utilization, the dominant contributing factorscomorbid disease burden or presepsis utilizationwere present prior to sepsis hospitalization. Early readmission patterns in sepsis appeared similar to those seen in other important hospital conditions, suggesting a role for shared posthospital, rather than just postsepsis, care approaches.
Disclosures
The funding for this study was provided by The Permanente Medical Group, Inc. and Kaiser Foundation Hospitals. The authors have no conflict of interests to disclose relevant to this article.
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Sepsis, the systemic inflammatory response to infection, is a major public health concern.[1] Worldwide, sepsis affects millions of hospitalized patients each year.[2] In the United States, it is the single most expensive cause of hospitalization.[3, 4, 5, 6] Multiple studies suggest that sepsis hospitalizations are also increasing in frequency.[3, 6, 7, 8, 9, 10]
Improved sepsis care has dramatically reduced in‐hospital mortality.[11, 12, 13] However, the result is a growing number of sepsis survivors discharged with new disability.[1, 9, 14, 15, 16] Despite being a common cause of hospitalization, little is known about how to improve postsepsis care.[15, 17, 18, 19] This contrasts with other, often less common, hospital conditions for which many studies evaluating readmission and postdischarge care are available.[20, 21, 22, 23] Identifying the factors contributing to high utilization could lend critical insight to designing interventions that improve long‐term sepsis outcomes.[24]
We conducted a retrospective study of sepsis patients discharged in 2010 at Kaiser Permanente Northern California (KPNC) to describe their posthospital trajectories. In this diverse community‐hospitalbased population, we sought to identify the patient‐level factors that impact the posthospital healthcare utilization of sepsis survivors.
METHODS
This study was approved by the KPNC institutional review board.
Setting
We conducted a retrospective study of sepsis patients aged 18 years admitted to KPNC hospitals in 2010 whose hospitalizations included an overnight stay, began in a KPNC hospital, and was not for peripartum care. We identified sepsis based on International Classification of Disease, 9th Edition principal diagnosis codes used at KPNC, which capture a similar population to that from the Angus definition (see Supporting Appendix, Table 1, in the online version of this article).[7, 25, 26] We denoted each patient's first sepsis hospitalization as the index event.
Predicted Hospital Mortality Quartiles (n=1,586 for Each Group) | |||||
---|---|---|---|---|---|
Overall | 1 | 2 | 3 | 4 | |
| |||||
Baseline | |||||
Age, y, mean | 71.915.7 | 62.317.8 | 71.214.2 | 75.612.7 | 78.612.2 |
By age category | |||||
<45 years | 410 (6.5) | 290 (18.3) | 71 (4.5) | 25 (1.6) | 24 (1.5) |
4564 years | 1,425 (22.5) | 539 (34.0) | 407 (25.7) | 292 (18.4) | 187 (11.8) |
6584 years | 3,036 (47.9) | 601 (37.9) | 814 (51.3) | 832 (52.5) | 789 (49.8) |
85 years | 1,473 (23.2) | 156 (9.8) | 294 (18.5) | 437 (27.6) | 586 (37.0) |
Male | 2,973 (46.9) | 686 (43.3) | 792 (49.9) | 750 (47.3) | 745 (47.0) |
Comorbidity | |||||
COPS2 score | 5143 | 2627 | 5441 | 6445 | 6245 |
Charlson score | 2.01.5 | 1.31.2 | 2.11.4 | 2.41.5 | 2.41.5 |
Hospitalization | |||||
LAPS2 severity score | 10742 | 6621 | 9020 | 11423 | 15928 |
Admitted via emergency department | 6,176 (97.4) | 1,522 (96.0) | 1,537 (96.9) | 1,539 (97.0) | 1,578 (99.5) |
Direct ICU admission | 1,730 (27.3) | 169 (10.7) | 309 (19.5) | 482 (30.4) | 770 (48.6) |
ICU transfer, at any time | 2,206 (34.8) | 279 (17.6) | 474 (29.9) | 603 (38.0) | 850 (53.6) |
Hospital mortality | |||||
Predicted, % | 10.513.8 | 1.00.1 | 3.40.1 | 8.32.3 | 29.415.8 |
Observed | 865 (13.6) | 26 (1.6) | 86 (5.4) | 197 (12.4) | 556 (35.1) |
Hospital length of stay, d | 5.86.4 | 4.43.8 | 5.45.7 | 6.68.0 | 6.66.9 |
We linked hospital episodes with existing KPNC inpatient databases to describe patient characteristics.[27, 28, 29, 30] We categorized patients by age (45, 4564, 6584, and 85 years) and used Charlson comorbidity scores and Comorbidity Point Scores 2 (COPS2) to quantify comorbid illness burden.[28, 30, 31, 32] We quantified acute severity of illness using the Laboratory Acute Physiology Scores 2 (LAPS2), which incorporates 15 laboratory values, 5 vital signs, and mental status prior to hospital admission (including emergency department data).[30] Both the COPS2 and LAPS2 are independently associated with hospital mortality.[30, 31] We also generated a summary predicted risk of hospital mortality based on a validated risk model and stratified patients by quartiles.[30] We determined whether patients were admitted to the intensive care unit (ICU).[29]
Outcomes
We used patients' health insurance administrative data to quantify postsepsis utilization. Within the KPNC integrated healthcare delivery system, uniform information systems capture all healthcare utilization of insured members including services received at non‐KPNC facilities.[28, 30] We collected utilization data from the year preceding index hospitalization (presepsis) and for the year after discharge date or until death (postsepsis). We ascertained mortality after discharge from KPNC medical records as well as state and national death record files.
We grouped services into facility‐based or outpatient categories. Facility‐based services included inpatient admission, subacute nursing facility or long‐term acute care, and emergency department visits. We grouped outpatient services as hospice, home health, outpatient surgery, clinic, or other (eg, laboratory). We excluded patients whose utilization records were not available over the full presepsis interval. Among these 1211 patients (12.5% of total), the median length of records prior to index hospitalization was 67 days, with a mean value of 117 days.
Statistical Analysis
Our primary outcomes of interest were hospital readmission and utilization in the year after sepsis. We defined a hospital readmission as any inpatient stay after the index hospitalization grouped within 1‐, 3‐, 6‐, and 12‐month intervals. We designated those within 30 days as an early readmission. We grouped readmission principal diagnoses, where available, by the 17 Healthcare Cost and Utilization Project (HCUP) Clinical Classifications Software multilevel categories with sepsis in the infectious category.[33, 34] In secondary analysis, we also designated other infectious diagnoses not included in the standard HCUP infection category (eg, pneumonia, meningitis, cellulitis) as infection (see Supporting Appendix in the online version of this article).
We quantified outpatient utilization based on the number of episodes recorded. For facility‐based utilization, we calculated patient length of stay intervals. Because patients surviving their index hospitalization might not survive the entire year after discharge, we also calculated utilization adjusted for patients' living days by dividing the total facility length of stay by the number of living days after discharge.
Continuous data are represented as mean (standard deviation [SD]) and categorical data as number (%). We compared groups with analysis of variance or 2 testing. We estimated survival with Kaplan‐Meier analysis (95% confidence interval) and compared groups with log‐rank testing. We compared pre‐ and postsepsis healthcare utilization with paired t tests.
To identify factors associated with early readmission after sepsis, we used a competing risks regression model.[35] The dependent variable was time to readmission and the competing hazard was death within 30 days without early readmission; patients without early readmission or death were censored at 30 days. The independent variables included age, gender, comorbid disease burden (COPS2), acute severity of illness (LAPS2), any use of intensive care, total index length of stay, and percentage of living days prior to sepsis hospitalization spent utilizing facility‐based care. We also used logistic regression to quantify the association between these variables and high postsepsis utilization; we defined high utilization as 15% of living days postsepsis spent in facility‐based care. For each model, we quantified the relative contribution of each predictor variable to model performance based on differences in log likelihoods.[35, 36] We conducted analyses using STATA/SE version 11.2 (StataCorp, College Station, TX) and considered a P value of <0.05 to be significant.
RESULTS
Cohort Characteristics
Our study cohort included 6344 patients with index sepsis hospitalizations in 2010 (Table 1). Mean age was 72 (SD 16) years including 1835 (28.9%) patients aged <65 years. During index hospitalizations, higher predicted mortality was associated with increased age, comorbid disease burden, and severity of illness (P<0.01 for each). ICU utilization increased across predicted mortality strata; for example, 10.7% of patients in the lowest quartile were admitted directly to the ICU compared with 48.6% in the highest quartile. In the highest quartile, observed mortality was 35.1%.
One‐Year Survival
A total of 5479 (86.4%) patients survived their index sepsis hospitalization. Overall survival after living discharge was 90.5% (range, 89.6%91.2%) at 30 days and 71.3% (range, 70.1%72.5%) at 1 year. However, postsepsis survival was strongly modified by age (Figure 1). For example, 1‐year survival was 94.1% (range, 91.2%96.0%) for <45 year olds and 54.4% (range, 51.5%57.2%) for 85 year olds (P<0.01). Survival was also modified by predicted mortality, however, not by ICU admission during index hospitalization (P=0.18) (see Supporting Appendix, Figure 1, in the online version of this article).
Hospital Readmission
Overall, 978 (17.9%) patients had early readmission after index discharge (Table 2); nearly half were readmitted at least once in the year following discharge. Rehospitalization frequency was slightly lower when including patients with incomplete presepsis data (see Supporting Appendix, Table 2, in the online version of this article). The frequency of hospital readmission varied based on patient age and severity of illness. For example, 22.3% of patients in the highest predicted mortality quartile had early readmission compared with 11.6% in the lowest. The median time from discharge to early readmission was 11 days. Principal diagnoses were available for 78.6% of all readmissions (see Supporting Appendix, Table 3, in the online version of this article). Between 28.3% and 42.7% of those readmissions were for infectious diagnoses (including sepsis).
Predicted Mortality Quartile | |||||
---|---|---|---|---|---|
Readmission | Overall | 1 | 2 | 3 | 4 |
Within 30 days | 978 (17.9) | 158 (11.6) | 242 (17.7) | 274 (20.0) | 304 (22.3) |
Within 90 days | 1,643 (30.1) | 276 (20.2) | 421 (30.8) | 463 (33.9) | 483 (35.4) |
Within 180 days | 2,061 (37.7) | 368 (26.9) | 540 (39.5) | 584 (42.7) | 569 (41.7) |
Within 365 days | 2,618 (47.9) | 498 (36.4) | 712 (52.1) | 723 (52.9) | 685 (50.2) |
Variable | Hazard Ratio for Early Readmission | Odds Ratio for High Utilization | ||
---|---|---|---|---|
HR (95% CI) | Relative Contribution | OR (95% CI) | Relative Contribution | |
| ||||
Age category | 1.2% | 11.1% | ||
<45 years | 1.00 [reference] | 1.00 [reference] | ||
4564 years | 0.86 (0.64‐1.16) | 2.22 (1.30‐3.83)a | ||
6584 years | 0.92 (0.69‐1.21) | 3.66 (2.17‐6.18)a | ||
85 years | 0.95 (0.70‐1.28) | 4.98 (2.92‐8.50)a | ||
Male | 0.99 (0.88‐1.13) | 0.0% | 0.86 (0.74‐1.00) | 0.1% |
Severity of illness (LAPS2) | 1.08 (1.04‐1.12)a | 12.4% | 1.22 (1.17‐1.27)a | 11.3% |
Comorbid illness (COPS2) | 1.16 (1.12‐1.19)a | 73.9% | 1.13 (1.09‐1.17)a | 5.9% |
Intensive care | 1.21 (1.05‐1.40)a | 5.2% | 1.02 (0.85‐1.21) | 0.0% |
Hospital length of stay, day | 1.01 (1.001.02)b | 6.6% | 1.04 (1.03‐1.06)a | 6.9% |
Prior utilization, per 10% | 0.98 (0.95‐1.02) | 0.7% | 1.74 (1.61‐1.88)a | 64.2% |
Healthcare Utilization
The unadjusted difference between pre‐ and postsepsis healthcare utilization among survivors was statistically significant for most categories but of modest clinical significance (see Supporting Appendix, Table 4, in the online version of this article). For example, the mean number of presepsis hospitalizations was 0.9 (1.4) compared to 1.0 (1.5) postsepsis (P<0.01). After adjusting for postsepsis living days, the difference in utilization was more pronounced (Figure 2). Overall, there was roughly a 3‐fold increase in the mean percentage of living days spent in facility‐based care between patients' pre‐ and postsepsis phases (5.3% vs 15.0%, P<0.01). Again, the difference was strongly modified by age. For patients aged <45 years, the difference was not statistically significant (2.4% vs 2.9%, P=0.32), whereas for those aged 65 years, it was highly significant (6.2% vs 18.5%, P<0.01).
Factors associated with early readmission included severity of illness, comorbid disease burden, index hospital length of stay, and intensive care (Table 3). However, the dominant factor explaining variation in the risk of early readmission was patients' prior comorbid disease burden (73.9%), followed by acute severity of illness (12.4%), total hospital length of stay (6.6%), and the need for intensive care (5.2%). Severity of illness and age were also significantly associated with higher odds of high postsepsis utilization; however, the dominant factor contributing to this risk was a history of high presepsis utilization (64.2%).
DISCUSSION
In this population‐based study in a community healthcare system, the impact of sepsis extended well beyond the initial hospitalization. One in 6 sepsis survivors was readmitted within 30 days, and roughly half were readmitted within 1 year. Fewer than half of rehospitalizations were for sepsis. Patients had a 3‐fold increase in the percentage of living days spent in hospitals or care facilities after sepsis hospitalization. Although age and acute severity of illness strongly modified healthcare utilization and mortality after sepsis, the dominant factors contributing to early readmission and high utilization ratescomorbid disease burden and presepsis healthcare utilizationwere present prior to hospitalization.
Sepsis is the single most expensive cause of US hospitalizations.[3, 4, 5] Despite its prevalence, there are little contemporary data identifying factors that impact healthcare utilization among sepsis survivors.[9, 16, 17, 19, 24, 36, 37] Recently, Prescott and others found that in Medicare beneficiaries, following severe sepsis, healthcare utilization was markedly increased.[17] More than one‐quarter of survivors were readmitted within 30 days, and 63.8% were readmitted within a year. Severe sepsis survivors also spent an average of 26% of their living days in a healthcare facility, a nearly 4‐fold increase compared to their presepsis phase. The current study included a population with a broader age and severity range; however, in a similar subgroup of patients, for those aged 65 years within the highest predicted mortality quartile, the frequency of readmission was similar. These findings are concordant with those from prior studies.[17, 19, 36, 37]
Among sepsis survivors, most readmissions were not for sepsis or infectious diagnoses, which is a novel finding with implications for designing approaches to reduce rehospitalization. The pattern in sepsis is similar to that seen in other common and costly hospital conditions.[17, 20, 23, 38, 39, 40] For example, between 18% and 25% of Medicare beneficiaries hospitalized for heart failure, acute myocardial infarction, or pneumonia were readmitted within 30 days; fewer than one‐third had the same diagnosis.[20] The timing of readmission in our sepsis cohort was also similar to that seen in other conditions.[20] For example, the median time of early readmission in this study was 11 days; it was between 10 and 12 days for patients with heart failure, pneumonia, and myocardial infarction.[20]
Krumholz and others suggest that the pattern of early rehospitalization after common acute conditions reflects a posthospital syndromean acquired, transient period of vulnerabilitythat could be the byproduct of common hospital factors.[20, 41] Such universal impairments might result from new physical and neurocognitive disability, nutritional deficiency, and sleep deprivation or delirium, among others.[41] If this construct were also true in sepsis, it could have important implications on the design of postsepsis care. However, prior studies suggest that sepsis patients may be particularly vulnerable to the sequelae of hospitalization.[2, 42, 43, 44, 45]
Among Medicare beneficiaries, Iwashyna and others reported that hospitalizations for severe sepsis resulted in significant increases in physical limitations and moderate to severe cognitive impairment.[1, 14, 46] Encephalopathy, sleep deprivation, and delirium are also frequently seen in sepsis patients.[47, 48] Furthermore, sepsis patients frequently need intensive care, which is also associated with increased patient disability and injury.[16, 46, 49, 50] We found that severity of illness and the need for intensive care were both predictive of the need for early readmission following sepsis. We also confirmed the results of prior studies suggesting that sepsis outcomes are strongly modified by age.[16, 19, 43, 51]
However, we found that the dominant factors contributing to patients' health trajectories were conditions present prior to admission. This finding is in accord with prior suggestions that acute severity of illness only partially predicts patients facing adverse posthospital sequelae.[23, 41, 52] Among sepsis patients, prior work demonstrates that inadequate consideration for presepsis level of function and utilization can result in an overestimation of the impact of sepsis on postdischarge health.[52, 53] Further, we found that the need for intensive care was not independently associated with an increased risk of high postsepsis utilization after adjusting for illness severity, a finding also seen in prior studies.[17, 23, 38, 51]
Taken together, our findings might suggest that an optimal approach to posthospital care in sepsis should focus on treatment approaches that address disease‐specific problems within the much larger context of common hospital risks. However, further study is necessary to clearly define the mechanisms by which age, severity of illness, and intensive care affect subsequent healthcare utilization. Furthermore, sepsis patients are a heterogeneous population in terms of severity of illness, site and pathogen of infection, and underlying comorbidity whose posthospital course remains incompletely characterized, limiting our ability to draw strong inferences.
These results should be interpreted in light of the study's limitations. First, our cohort included patients with healthcare insurance within a community‐based healthcare system. Care within the KPNC system, which bears similarities with accountable care organizations, is enhanced through service integration and a comprehensive health information system. Although prior studies suggest that these characteristics result in improved population‐based care, it is unclear whether there is a similar impact in hospital‐based conditions such as sepsis.[54, 55] Furthermore, care within an integrated system may impact posthospital utilization patterns and could limit generalizability. However, prior studies demonstrate the similarity of KPNC members to other patients in the same region in terms of age, socioeconomics, overall health behaviors, and racial/ethnic diversity.[56] Second, our study did not characterize organ dysfunction based on diagnosis coding, a common feature of sepsis studies that lack detailed physiologic severity data.[4, 5, 6, 8, 26] Instead, we focused on using granular laboratory and vital signs data to ensure accurate risk adjustment using a validated system developed in >400,000 hospitalizations.[30] Although this method may hamper comparisons with existing studies, traditional methods of grading severity by diagnosis codes can be vulnerable to biases resulting in wide variability.[10, 23, 26, 57, 58] Nonetheless, it is likely that characterizing preexisting and acute organ dysfunction will improve risk stratification in the heterogeneous sepsis population. Third, this study did not include data regarding patients' functional status, which has been shown to strongly predict patient outcomes following hospitalization. Fourth, this study did not address the cost of care following sepsis hospitalizations.[19, 59] Finally, our study excluded patients with incomplete utilization records, a choice designed to avoid the spurious inferences that can result from such comparisons.[53]
In summary, we found that sepsis exacted a considerable toll on patients in the hospital and in the year following discharge. Sepsis patients were frequently rehospitalized within a month of discharge, and on average had a 3‐fold increase in their subsequent time spent in healthcare facilities. Although age, severity of illness, and the need for ICU care impacted postsepsis utilization, the dominant contributing factorscomorbid disease burden or presepsis utilizationwere present prior to sepsis hospitalization. Early readmission patterns in sepsis appeared similar to those seen in other important hospital conditions, suggesting a role for shared posthospital, rather than just postsepsis, care approaches.
Disclosures
The funding for this study was provided by The Permanente Medical Group, Inc. and Kaiser Foundation Hospitals. The authors have no conflict of interests to disclose relevant to this article.
Sepsis, the systemic inflammatory response to infection, is a major public health concern.[1] Worldwide, sepsis affects millions of hospitalized patients each year.[2] In the United States, it is the single most expensive cause of hospitalization.[3, 4, 5, 6] Multiple studies suggest that sepsis hospitalizations are also increasing in frequency.[3, 6, 7, 8, 9, 10]
Improved sepsis care has dramatically reduced in‐hospital mortality.[11, 12, 13] However, the result is a growing number of sepsis survivors discharged with new disability.[1, 9, 14, 15, 16] Despite being a common cause of hospitalization, little is known about how to improve postsepsis care.[15, 17, 18, 19] This contrasts with other, often less common, hospital conditions for which many studies evaluating readmission and postdischarge care are available.[20, 21, 22, 23] Identifying the factors contributing to high utilization could lend critical insight to designing interventions that improve long‐term sepsis outcomes.[24]
We conducted a retrospective study of sepsis patients discharged in 2010 at Kaiser Permanente Northern California (KPNC) to describe their posthospital trajectories. In this diverse community‐hospitalbased population, we sought to identify the patient‐level factors that impact the posthospital healthcare utilization of sepsis survivors.
METHODS
This study was approved by the KPNC institutional review board.
Setting
We conducted a retrospective study of sepsis patients aged 18 years admitted to KPNC hospitals in 2010 whose hospitalizations included an overnight stay, began in a KPNC hospital, and was not for peripartum care. We identified sepsis based on International Classification of Disease, 9th Edition principal diagnosis codes used at KPNC, which capture a similar population to that from the Angus definition (see Supporting Appendix, Table 1, in the online version of this article).[7, 25, 26] We denoted each patient's first sepsis hospitalization as the index event.
Predicted Hospital Mortality Quartiles (n=1,586 for Each Group) | |||||
---|---|---|---|---|---|
Overall | 1 | 2 | 3 | 4 | |
| |||||
Baseline | |||||
Age, y, mean | 71.915.7 | 62.317.8 | 71.214.2 | 75.612.7 | 78.612.2 |
By age category | |||||
<45 years | 410 (6.5) | 290 (18.3) | 71 (4.5) | 25 (1.6) | 24 (1.5) |
4564 years | 1,425 (22.5) | 539 (34.0) | 407 (25.7) | 292 (18.4) | 187 (11.8) |
6584 years | 3,036 (47.9) | 601 (37.9) | 814 (51.3) | 832 (52.5) | 789 (49.8) |
85 years | 1,473 (23.2) | 156 (9.8) | 294 (18.5) | 437 (27.6) | 586 (37.0) |
Male | 2,973 (46.9) | 686 (43.3) | 792 (49.9) | 750 (47.3) | 745 (47.0) |
Comorbidity | |||||
COPS2 score | 5143 | 2627 | 5441 | 6445 | 6245 |
Charlson score | 2.01.5 | 1.31.2 | 2.11.4 | 2.41.5 | 2.41.5 |
Hospitalization | |||||
LAPS2 severity score | 10742 | 6621 | 9020 | 11423 | 15928 |
Admitted via emergency department | 6,176 (97.4) | 1,522 (96.0) | 1,537 (96.9) | 1,539 (97.0) | 1,578 (99.5) |
Direct ICU admission | 1,730 (27.3) | 169 (10.7) | 309 (19.5) | 482 (30.4) | 770 (48.6) |
ICU transfer, at any time | 2,206 (34.8) | 279 (17.6) | 474 (29.9) | 603 (38.0) | 850 (53.6) |
Hospital mortality | |||||
Predicted, % | 10.513.8 | 1.00.1 | 3.40.1 | 8.32.3 | 29.415.8 |
Observed | 865 (13.6) | 26 (1.6) | 86 (5.4) | 197 (12.4) | 556 (35.1) |
Hospital length of stay, d | 5.86.4 | 4.43.8 | 5.45.7 | 6.68.0 | 6.66.9 |
We linked hospital episodes with existing KPNC inpatient databases to describe patient characteristics.[27, 28, 29, 30] We categorized patients by age (45, 4564, 6584, and 85 years) and used Charlson comorbidity scores and Comorbidity Point Scores 2 (COPS2) to quantify comorbid illness burden.[28, 30, 31, 32] We quantified acute severity of illness using the Laboratory Acute Physiology Scores 2 (LAPS2), which incorporates 15 laboratory values, 5 vital signs, and mental status prior to hospital admission (including emergency department data).[30] Both the COPS2 and LAPS2 are independently associated with hospital mortality.[30, 31] We also generated a summary predicted risk of hospital mortality based on a validated risk model and stratified patients by quartiles.[30] We determined whether patients were admitted to the intensive care unit (ICU).[29]
Outcomes
We used patients' health insurance administrative data to quantify postsepsis utilization. Within the KPNC integrated healthcare delivery system, uniform information systems capture all healthcare utilization of insured members including services received at non‐KPNC facilities.[28, 30] We collected utilization data from the year preceding index hospitalization (presepsis) and for the year after discharge date or until death (postsepsis). We ascertained mortality after discharge from KPNC medical records as well as state and national death record files.
We grouped services into facility‐based or outpatient categories. Facility‐based services included inpatient admission, subacute nursing facility or long‐term acute care, and emergency department visits. We grouped outpatient services as hospice, home health, outpatient surgery, clinic, or other (eg, laboratory). We excluded patients whose utilization records were not available over the full presepsis interval. Among these 1211 patients (12.5% of total), the median length of records prior to index hospitalization was 67 days, with a mean value of 117 days.
Statistical Analysis
Our primary outcomes of interest were hospital readmission and utilization in the year after sepsis. We defined a hospital readmission as any inpatient stay after the index hospitalization grouped within 1‐, 3‐, 6‐, and 12‐month intervals. We designated those within 30 days as an early readmission. We grouped readmission principal diagnoses, where available, by the 17 Healthcare Cost and Utilization Project (HCUP) Clinical Classifications Software multilevel categories with sepsis in the infectious category.[33, 34] In secondary analysis, we also designated other infectious diagnoses not included in the standard HCUP infection category (eg, pneumonia, meningitis, cellulitis) as infection (see Supporting Appendix in the online version of this article).
We quantified outpatient utilization based on the number of episodes recorded. For facility‐based utilization, we calculated patient length of stay intervals. Because patients surviving their index hospitalization might not survive the entire year after discharge, we also calculated utilization adjusted for patients' living days by dividing the total facility length of stay by the number of living days after discharge.
Continuous data are represented as mean (standard deviation [SD]) and categorical data as number (%). We compared groups with analysis of variance or 2 testing. We estimated survival with Kaplan‐Meier analysis (95% confidence interval) and compared groups with log‐rank testing. We compared pre‐ and postsepsis healthcare utilization with paired t tests.
To identify factors associated with early readmission after sepsis, we used a competing risks regression model.[35] The dependent variable was time to readmission and the competing hazard was death within 30 days without early readmission; patients without early readmission or death were censored at 30 days. The independent variables included age, gender, comorbid disease burden (COPS2), acute severity of illness (LAPS2), any use of intensive care, total index length of stay, and percentage of living days prior to sepsis hospitalization spent utilizing facility‐based care. We also used logistic regression to quantify the association between these variables and high postsepsis utilization; we defined high utilization as 15% of living days postsepsis spent in facility‐based care. For each model, we quantified the relative contribution of each predictor variable to model performance based on differences in log likelihoods.[35, 36] We conducted analyses using STATA/SE version 11.2 (StataCorp, College Station, TX) and considered a P value of <0.05 to be significant.
RESULTS
Cohort Characteristics
Our study cohort included 6344 patients with index sepsis hospitalizations in 2010 (Table 1). Mean age was 72 (SD 16) years including 1835 (28.9%) patients aged <65 years. During index hospitalizations, higher predicted mortality was associated with increased age, comorbid disease burden, and severity of illness (P<0.01 for each). ICU utilization increased across predicted mortality strata; for example, 10.7% of patients in the lowest quartile were admitted directly to the ICU compared with 48.6% in the highest quartile. In the highest quartile, observed mortality was 35.1%.
One‐Year Survival
A total of 5479 (86.4%) patients survived their index sepsis hospitalization. Overall survival after living discharge was 90.5% (range, 89.6%91.2%) at 30 days and 71.3% (range, 70.1%72.5%) at 1 year. However, postsepsis survival was strongly modified by age (Figure 1). For example, 1‐year survival was 94.1% (range, 91.2%96.0%) for <45 year olds and 54.4% (range, 51.5%57.2%) for 85 year olds (P<0.01). Survival was also modified by predicted mortality, however, not by ICU admission during index hospitalization (P=0.18) (see Supporting Appendix, Figure 1, in the online version of this article).
Hospital Readmission
Overall, 978 (17.9%) patients had early readmission after index discharge (Table 2); nearly half were readmitted at least once in the year following discharge. Rehospitalization frequency was slightly lower when including patients with incomplete presepsis data (see Supporting Appendix, Table 2, in the online version of this article). The frequency of hospital readmission varied based on patient age and severity of illness. For example, 22.3% of patients in the highest predicted mortality quartile had early readmission compared with 11.6% in the lowest. The median time from discharge to early readmission was 11 days. Principal diagnoses were available for 78.6% of all readmissions (see Supporting Appendix, Table 3, in the online version of this article). Between 28.3% and 42.7% of those readmissions were for infectious diagnoses (including sepsis).
Predicted Mortality Quartile | |||||
---|---|---|---|---|---|
Readmission | Overall | 1 | 2 | 3 | 4 |
Within 30 days | 978 (17.9) | 158 (11.6) | 242 (17.7) | 274 (20.0) | 304 (22.3) |
Within 90 days | 1,643 (30.1) | 276 (20.2) | 421 (30.8) | 463 (33.9) | 483 (35.4) |
Within 180 days | 2,061 (37.7) | 368 (26.9) | 540 (39.5) | 584 (42.7) | 569 (41.7) |
Within 365 days | 2,618 (47.9) | 498 (36.4) | 712 (52.1) | 723 (52.9) | 685 (50.2) |
Variable | Hazard Ratio for Early Readmission | Odds Ratio for High Utilization | ||
---|---|---|---|---|
HR (95% CI) | Relative Contribution | OR (95% CI) | Relative Contribution | |
| ||||
Age category | 1.2% | 11.1% | ||
<45 years | 1.00 [reference] | 1.00 [reference] | ||
4564 years | 0.86 (0.64‐1.16) | 2.22 (1.30‐3.83)a | ||
6584 years | 0.92 (0.69‐1.21) | 3.66 (2.17‐6.18)a | ||
85 years | 0.95 (0.70‐1.28) | 4.98 (2.92‐8.50)a | ||
Male | 0.99 (0.88‐1.13) | 0.0% | 0.86 (0.74‐1.00) | 0.1% |
Severity of illness (LAPS2) | 1.08 (1.04‐1.12)a | 12.4% | 1.22 (1.17‐1.27)a | 11.3% |
Comorbid illness (COPS2) | 1.16 (1.12‐1.19)a | 73.9% | 1.13 (1.09‐1.17)a | 5.9% |
Intensive care | 1.21 (1.05‐1.40)a | 5.2% | 1.02 (0.85‐1.21) | 0.0% |
Hospital length of stay, day | 1.01 (1.001.02)b | 6.6% | 1.04 (1.03‐1.06)a | 6.9% |
Prior utilization, per 10% | 0.98 (0.95‐1.02) | 0.7% | 1.74 (1.61‐1.88)a | 64.2% |
Healthcare Utilization
The unadjusted difference between pre‐ and postsepsis healthcare utilization among survivors was statistically significant for most categories but of modest clinical significance (see Supporting Appendix, Table 4, in the online version of this article). For example, the mean number of presepsis hospitalizations was 0.9 (1.4) compared to 1.0 (1.5) postsepsis (P<0.01). After adjusting for postsepsis living days, the difference in utilization was more pronounced (Figure 2). Overall, there was roughly a 3‐fold increase in the mean percentage of living days spent in facility‐based care between patients' pre‐ and postsepsis phases (5.3% vs 15.0%, P<0.01). Again, the difference was strongly modified by age. For patients aged <45 years, the difference was not statistically significant (2.4% vs 2.9%, P=0.32), whereas for those aged 65 years, it was highly significant (6.2% vs 18.5%, P<0.01).
Factors associated with early readmission included severity of illness, comorbid disease burden, index hospital length of stay, and intensive care (Table 3). However, the dominant factor explaining variation in the risk of early readmission was patients' prior comorbid disease burden (73.9%), followed by acute severity of illness (12.4%), total hospital length of stay (6.6%), and the need for intensive care (5.2%). Severity of illness and age were also significantly associated with higher odds of high postsepsis utilization; however, the dominant factor contributing to this risk was a history of high presepsis utilization (64.2%).
DISCUSSION
In this population‐based study in a community healthcare system, the impact of sepsis extended well beyond the initial hospitalization. One in 6 sepsis survivors was readmitted within 30 days, and roughly half were readmitted within 1 year. Fewer than half of rehospitalizations were for sepsis. Patients had a 3‐fold increase in the percentage of living days spent in hospitals or care facilities after sepsis hospitalization. Although age and acute severity of illness strongly modified healthcare utilization and mortality after sepsis, the dominant factors contributing to early readmission and high utilization ratescomorbid disease burden and presepsis healthcare utilizationwere present prior to hospitalization.
Sepsis is the single most expensive cause of US hospitalizations.[3, 4, 5] Despite its prevalence, there are little contemporary data identifying factors that impact healthcare utilization among sepsis survivors.[9, 16, 17, 19, 24, 36, 37] Recently, Prescott and others found that in Medicare beneficiaries, following severe sepsis, healthcare utilization was markedly increased.[17] More than one‐quarter of survivors were readmitted within 30 days, and 63.8% were readmitted within a year. Severe sepsis survivors also spent an average of 26% of their living days in a healthcare facility, a nearly 4‐fold increase compared to their presepsis phase. The current study included a population with a broader age and severity range; however, in a similar subgroup of patients, for those aged 65 years within the highest predicted mortality quartile, the frequency of readmission was similar. These findings are concordant with those from prior studies.[17, 19, 36, 37]
Among sepsis survivors, most readmissions were not for sepsis or infectious diagnoses, which is a novel finding with implications for designing approaches to reduce rehospitalization. The pattern in sepsis is similar to that seen in other common and costly hospital conditions.[17, 20, 23, 38, 39, 40] For example, between 18% and 25% of Medicare beneficiaries hospitalized for heart failure, acute myocardial infarction, or pneumonia were readmitted within 30 days; fewer than one‐third had the same diagnosis.[20] The timing of readmission in our sepsis cohort was also similar to that seen in other conditions.[20] For example, the median time of early readmission in this study was 11 days; it was between 10 and 12 days for patients with heart failure, pneumonia, and myocardial infarction.[20]
Krumholz and others suggest that the pattern of early rehospitalization after common acute conditions reflects a posthospital syndromean acquired, transient period of vulnerabilitythat could be the byproduct of common hospital factors.[20, 41] Such universal impairments might result from new physical and neurocognitive disability, nutritional deficiency, and sleep deprivation or delirium, among others.[41] If this construct were also true in sepsis, it could have important implications on the design of postsepsis care. However, prior studies suggest that sepsis patients may be particularly vulnerable to the sequelae of hospitalization.[2, 42, 43, 44, 45]
Among Medicare beneficiaries, Iwashyna and others reported that hospitalizations for severe sepsis resulted in significant increases in physical limitations and moderate to severe cognitive impairment.[1, 14, 46] Encephalopathy, sleep deprivation, and delirium are also frequently seen in sepsis patients.[47, 48] Furthermore, sepsis patients frequently need intensive care, which is also associated with increased patient disability and injury.[16, 46, 49, 50] We found that severity of illness and the need for intensive care were both predictive of the need for early readmission following sepsis. We also confirmed the results of prior studies suggesting that sepsis outcomes are strongly modified by age.[16, 19, 43, 51]
However, we found that the dominant factors contributing to patients' health trajectories were conditions present prior to admission. This finding is in accord with prior suggestions that acute severity of illness only partially predicts patients facing adverse posthospital sequelae.[23, 41, 52] Among sepsis patients, prior work demonstrates that inadequate consideration for presepsis level of function and utilization can result in an overestimation of the impact of sepsis on postdischarge health.[52, 53] Further, we found that the need for intensive care was not independently associated with an increased risk of high postsepsis utilization after adjusting for illness severity, a finding also seen in prior studies.[17, 23, 38, 51]
Taken together, our findings might suggest that an optimal approach to posthospital care in sepsis should focus on treatment approaches that address disease‐specific problems within the much larger context of common hospital risks. However, further study is necessary to clearly define the mechanisms by which age, severity of illness, and intensive care affect subsequent healthcare utilization. Furthermore, sepsis patients are a heterogeneous population in terms of severity of illness, site and pathogen of infection, and underlying comorbidity whose posthospital course remains incompletely characterized, limiting our ability to draw strong inferences.
These results should be interpreted in light of the study's limitations. First, our cohort included patients with healthcare insurance within a community‐based healthcare system. Care within the KPNC system, which bears similarities with accountable care organizations, is enhanced through service integration and a comprehensive health information system. Although prior studies suggest that these characteristics result in improved population‐based care, it is unclear whether there is a similar impact in hospital‐based conditions such as sepsis.[54, 55] Furthermore, care within an integrated system may impact posthospital utilization patterns and could limit generalizability. However, prior studies demonstrate the similarity of KPNC members to other patients in the same region in terms of age, socioeconomics, overall health behaviors, and racial/ethnic diversity.[56] Second, our study did not characterize organ dysfunction based on diagnosis coding, a common feature of sepsis studies that lack detailed physiologic severity data.[4, 5, 6, 8, 26] Instead, we focused on using granular laboratory and vital signs data to ensure accurate risk adjustment using a validated system developed in >400,000 hospitalizations.[30] Although this method may hamper comparisons with existing studies, traditional methods of grading severity by diagnosis codes can be vulnerable to biases resulting in wide variability.[10, 23, 26, 57, 58] Nonetheless, it is likely that characterizing preexisting and acute organ dysfunction will improve risk stratification in the heterogeneous sepsis population. Third, this study did not include data regarding patients' functional status, which has been shown to strongly predict patient outcomes following hospitalization. Fourth, this study did not address the cost of care following sepsis hospitalizations.[19, 59] Finally, our study excluded patients with incomplete utilization records, a choice designed to avoid the spurious inferences that can result from such comparisons.[53]
In summary, we found that sepsis exacted a considerable toll on patients in the hospital and in the year following discharge. Sepsis patients were frequently rehospitalized within a month of discharge, and on average had a 3‐fold increase in their subsequent time spent in healthcare facilities. Although age, severity of illness, and the need for ICU care impacted postsepsis utilization, the dominant contributing factorscomorbid disease burden or presepsis utilizationwere present prior to sepsis hospitalization. Early readmission patterns in sepsis appeared similar to those seen in other important hospital conditions, suggesting a role for shared posthospital, rather than just postsepsis, care approaches.
Disclosures
The funding for this study was provided by The Permanente Medical Group, Inc. and Kaiser Foundation Hospitals. The authors have no conflict of interests to disclose relevant to this article.
- The lingering consequences of sepsis: a hidden public health disaster? JAMA. 2010;304(16):1833–1834. .
- Surviving Sepsis Campaign Guidelines Committee including the Pediatric Subgroup. Surviving sepsis campaign: international guidelines for management of severe sepsis and septic shock: 2012. Crit Care Med. 2013;41(2):580–637. , , , et al.;
- Costs for hospital stays in the United States, 2010. HCUP statistical brief #16. January 2013. Rockville, MD: Agency for Healthcare Research and Quality; 2013. Available at: http://www.hcup‐us.ahrq.gov/reports/statbriefs/sb146.pdf. Accessed October 1, 2013. , , .
- The epidemiology of sepsis in the United States from 1979 through 2000. N Engl J Med. 2003;348(16):1546–1554. , , , .
- Epidemiology of severe sepsis in the United States: analysis of incidence, outcome, and associated costs of care. Crit Care Med. 2001;29(7):1303–1310. , , , , , .
- Rapid increase in hospitalization and mortality rates for severe sepsis in the United States: a trend analysis from 1993 to 2003. Crit Care Med. 2007;35(5):1244–1250. , , , .
- Septicemia in U.S. hospitals, 2009. HCUP statistical brief #122. October 2011. Rockville, MD: Agency for Healthcare Research and Quality; 2011. Available at: http://www.hcup‐us.ahrq.gov/reports/statbriefs/sb122.pdf. Accessed October 1, 2013. , , .
- Hospitalizations, costs, and outcomes of severe sepsis in the United States 2003 to 2007. Crit Care Med. 2012;40(3):754–761. , , , , , .
- Population burden of long‐term survivorship after severe sepsis in older Americans. J Am Geriatr Soc. 2012;60(6):1070–1077. , , , .
- Benchmarking the incidence and mortality of severe sepsis in the United States. Crit Care Med. 2013;41(5):1167–1174. , , , .
- Outcomes of the Surviving Sepsis Campaign in intensive care units in the USA and Europe: a prospective cohort study. Lancet Infect Dis. 2012;12(12):919–924. , , , et al.
- Reducing mortality in severe sepsis: the Surviving Sepsis Campaign. Clin Chest Med. 2008;29(4):721–733, x. , , , .
- Early Goal‐Directed Therapy Collaborative Group. Early goal‐directed therapy in the treatment of severe sepsis and septic shock. N Engl J Med. 2001;345(19):1368–1377. , , , et al.;
- Long‐term cognitive impairment and functional disability among survivors of severe sepsis. JAMA. 2010;304(16):1787–1794. , , , .
- Long‐term mortality and quality of life in sepsis: a systematic review. Crit Care Med. 2010;38(5):1276–1283. , , , , , .
- the Scottish Critical Care Trials Group and the Scottish Intensive Care Society Audit Group. Mortality and quality of life in the five years after severe sepsis. Crit Care. 2013;17(2):R70. , , , et al.;
- Post‐Discharge Health Care Use Is Markedly Higher in Survivors of Severe Sepsis. Am J Respir Crit Care Med 2013;187:A1573. , , , , .
- Long‐term survival and function after suspected gram‐negative sepsis. JAMA. 1995;274(4):338–345. , , , .
- Long‐term mortality and medical care charges in patients with severe sepsis. Crit Care Med. 2003;31(9):2316–2323. , , , , .
- Diagnoses and timing of 30‐day readmissions after hospitalization for heart failure, acute myocardial infarction, or pneumonia. JAMA. 2013;309(4):355–363. , , , et al.
- A systematic review and meta‐analysis of studies comparing readmission rates and mortality rates in patients with heart failure. Arch Intern Med. 2004;164(21):2315–2320. , , , , .
- Hospitalizations for heart failure in the United States—a sign of hope. JAMA. 2011;306(15):1705–1706. , .
- Risk prediction models for hospital readmission: a systematic review. JAMA. 2011;306(15):1688–1698. , , , et al.
- Sepsis after Scotland: enough with the averages, show us the effect modifiers. Crit Care. 2013;17(3):148. , .
- Kaiser Permanente's performance improvement system, part 3: multisite improvements in care for patients with sepsis. Jt Comm J Qual Patient Saf. 2011;37(11): 483–493. , , , et al.
- Identifying patients with severe sepsis using administrative claims: patient‐level validation of the Angus implementation of the International Consensus Conference Definition of Severe Sepsis [published online ahead of print September 18, 2012]. Med Care. doi: 10.1097/MLR.0b013e318268ac86. Epub ahead of print. , , , et al.
- Linking automated databases for research in managed care settings. Ann Intern Med. 1997;127(8 pt 2):719–724. .
- Risk‐adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , , , .
- An electronic Simplified Acute Physiology Score‐based risk adjustment score for critical illness in an integrated healthcare system. Crit Care Med. 2013;41(1):41–48. , , , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated health care delivery system. Med Care. 2013;51(5):446–453. , , , , .
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population. J Clin Epidemiol. 2009;63(7):798–803. , , , .
- Casemix adjustment of managed care claims data using the clinical classification for health policy research method. Med Care. 1998;36(7):1108–1113. , , , , , .
- Agency for Healthcare Research and Quality Healthcare Cost and Utilization Project. Clinical Classifications Software (CCS) for ICD‐9‐CM Fact Sheet. Available at: http://www.hcup‐us.ahrq.gov/toolssoftware/ccs/ccsfactsheet.jsp. Accessed January 20, 2013.
- A proportional hazards model for the subdistribution of a competing risk. J Am Stat Assoc. 1997;94(446):496–509. , .
- Severe sepsis in managed care: analysis of incidence, one‐year mortality, and associated costs of care. J Manag Care Pharm. 2004;10(6):521–530. , , .
- Detailed cost analysis of care for survivors of severe sepsis. Crit Care Med. 2004;32(4):981–985. , , , , , .
- Readmissions among patients with severe sepsis/septic shock among inner‐city minority New Yorkers. Chest. 2012;142:286A. , , , , et al.
- Readmission and late mortality after pediatric severe sepsis. Pediatrics. 2009;123(3):849–857. , , .
- Rehospitalizations among patients in the Medicare fee‐for‐service program. N Engl J Med. 2009;360(14):1418–1428. , , .
- Post‐hospital syndrome—an acquired, transient condition of generalized risk. N Engl J Med. 2013;368(2):100–102. .
- Definitions for sepsis and organ failure and guidelines for the use of innovative therapies in sepsis. The ACCP/SCCM Consensus Conference Committee. American College of Chest Physicians/Society of Critical Care Medicine. Chest. 1992;101(6):1644–1655. , , , et al.
- The effect of age on the development and outcome of adult sepsis. Crit Care Med. 2006;34(1):15–21. , , .
- Multiple systems organ failure: failure of host defense homeostasis. Crit Care Clin. 1989;5(2):199–220. , .
- Pathophysiology of sepsis. Am J Pathol. 2007;170(5):1435–1444. .
- Surviving intensive care: a report from the 2002 Brussels Roundtable. Intensive Care Med. 2003;29(3):368–377. , .
- The encephalopathy in sepsis. Crit Care Clin. 2008;24(1):67–82, viii. , , .
- Sepsis‐associated encephalopathy. Nat Rev Neurol. 2012;8(10):557–566. , .
- Improving long‐term outcomes after discharge from intensive care unit: report from a stakeholders' conference. Crit Care Med. 2012;40(2):502–509. , , , et al.
- The association between sepsis and potential medical injury among hospitalized patients. Chest. 2012;142(3):606–613. , , , , .
- Three‐year outcomes for Medicare beneficiaries who survive intensive care. JAMA. 2010;303(9):849–856. , , , , , .
- Does acute organ dysfunction predict patient‐centered outcomes? Chest. 2002;121(6):1963–1971. , , , , , .
- Spurious inferences about long‐term outcomes: the case of severe sepsis and geriatric conditions. Am J Respir Crit Care Med. 2012;185(8):835–841. , , , .
- Population trends in the incidence and outcomes of acute myocardial infarction. N Engl J Med. 2010;362(23):2155–2165. , , , , , .
- Outpatient electronic health records and the clinical care and outcomes of patients with diabetes mellitus. Ann Intern Med. 2012;157(7):482–489. , , , et al.,
- Similarity of the adult Kaiser Permanente membership in Northern California to the insured and general population in Northern California: statistics from the 2009 California Health Interview Survey. Internal Division of Research Report. Oakland, CA: Kaiser Permanente Division of Research; January 24, 2012. Available at: http://www.dor.kaiser.org/external/chis_non_kp_2009. Accessed January 20, 2013. .
- Association of diagnostic coding with trends in hospitalizations and mortality of patients with pneumonia, 2003–2009. JAMA. 2012;307(13):1405–1413. , , , , .
- Finding pure and simple truths with administrative data. JAMA. 2012;307(13):1433–1435. , .
- Cost savings attributable to reductions in intensive care unit length of stay for mechanically ventilated patients. Med Care. 2008;46(12):1226–1233. , , , .
- The lingering consequences of sepsis: a hidden public health disaster? JAMA. 2010;304(16):1833–1834. .
- Surviving Sepsis Campaign Guidelines Committee including the Pediatric Subgroup. Surviving sepsis campaign: international guidelines for management of severe sepsis and septic shock: 2012. Crit Care Med. 2013;41(2):580–637. , , , et al.;
- Costs for hospital stays in the United States, 2010. HCUP statistical brief #16. January 2013. Rockville, MD: Agency for Healthcare Research and Quality; 2013. Available at: http://www.hcup‐us.ahrq.gov/reports/statbriefs/sb146.pdf. Accessed October 1, 2013. , , .
- The epidemiology of sepsis in the United States from 1979 through 2000. N Engl J Med. 2003;348(16):1546–1554. , , , .
- Epidemiology of severe sepsis in the United States: analysis of incidence, outcome, and associated costs of care. Crit Care Med. 2001;29(7):1303–1310. , , , , , .
- Rapid increase in hospitalization and mortality rates for severe sepsis in the United States: a trend analysis from 1993 to 2003. Crit Care Med. 2007;35(5):1244–1250. , , , .
- Septicemia in U.S. hospitals, 2009. HCUP statistical brief #122. October 2011. Rockville, MD: Agency for Healthcare Research and Quality; 2011. Available at: http://www.hcup‐us.ahrq.gov/reports/statbriefs/sb122.pdf. Accessed October 1, 2013. , , .
- Hospitalizations, costs, and outcomes of severe sepsis in the United States 2003 to 2007. Crit Care Med. 2012;40(3):754–761. , , , , , .
- Population burden of long‐term survivorship after severe sepsis in older Americans. J Am Geriatr Soc. 2012;60(6):1070–1077. , , , .
- Benchmarking the incidence and mortality of severe sepsis in the United States. Crit Care Med. 2013;41(5):1167–1174. , , , .
- Outcomes of the Surviving Sepsis Campaign in intensive care units in the USA and Europe: a prospective cohort study. Lancet Infect Dis. 2012;12(12):919–924. , , , et al.
- Reducing mortality in severe sepsis: the Surviving Sepsis Campaign. Clin Chest Med. 2008;29(4):721–733, x. , , , .
- Early Goal‐Directed Therapy Collaborative Group. Early goal‐directed therapy in the treatment of severe sepsis and septic shock. N Engl J Med. 2001;345(19):1368–1377. , , , et al.;
- Long‐term cognitive impairment and functional disability among survivors of severe sepsis. JAMA. 2010;304(16):1787–1794. , , , .
- Long‐term mortality and quality of life in sepsis: a systematic review. Crit Care Med. 2010;38(5):1276–1283. , , , , , .
- the Scottish Critical Care Trials Group and the Scottish Intensive Care Society Audit Group. Mortality and quality of life in the five years after severe sepsis. Crit Care. 2013;17(2):R70. , , , et al.;
- Post‐Discharge Health Care Use Is Markedly Higher in Survivors of Severe Sepsis. Am J Respir Crit Care Med 2013;187:A1573. , , , , .
- Long‐term survival and function after suspected gram‐negative sepsis. JAMA. 1995;274(4):338–345. , , , .
- Long‐term mortality and medical care charges in patients with severe sepsis. Crit Care Med. 2003;31(9):2316–2323. , , , , .
- Diagnoses and timing of 30‐day readmissions after hospitalization for heart failure, acute myocardial infarction, or pneumonia. JAMA. 2013;309(4):355–363. , , , et al.
- A systematic review and meta‐analysis of studies comparing readmission rates and mortality rates in patients with heart failure. Arch Intern Med. 2004;164(21):2315–2320. , , , , .
- Hospitalizations for heart failure in the United States—a sign of hope. JAMA. 2011;306(15):1705–1706. , .
- Risk prediction models for hospital readmission: a systematic review. JAMA. 2011;306(15):1688–1698. , , , et al.
- Sepsis after Scotland: enough with the averages, show us the effect modifiers. Crit Care. 2013;17(3):148. , .
- Kaiser Permanente's performance improvement system, part 3: multisite improvements in care for patients with sepsis. Jt Comm J Qual Patient Saf. 2011;37(11): 483–493. , , , et al.
- Identifying patients with severe sepsis using administrative claims: patient‐level validation of the Angus implementation of the International Consensus Conference Definition of Severe Sepsis [published online ahead of print September 18, 2012]. Med Care. doi: 10.1097/MLR.0b013e318268ac86. Epub ahead of print. , , , et al.
- Linking automated databases for research in managed care settings. Ann Intern Med. 1997;127(8 pt 2):719–724. .
- Risk‐adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , , , .
- An electronic Simplified Acute Physiology Score‐based risk adjustment score for critical illness in an integrated healthcare system. Crit Care Med. 2013;41(1):41–48. , , , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated health care delivery system. Med Care. 2013;51(5):446–453. , , , , .
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population. J Clin Epidemiol. 2009;63(7):798–803. , , , .
- Casemix adjustment of managed care claims data using the clinical classification for health policy research method. Med Care. 1998;36(7):1108–1113. , , , , , .
- Agency for Healthcare Research and Quality Healthcare Cost and Utilization Project. Clinical Classifications Software (CCS) for ICD‐9‐CM Fact Sheet. Available at: http://www.hcup‐us.ahrq.gov/toolssoftware/ccs/ccsfactsheet.jsp. Accessed January 20, 2013.
- A proportional hazards model for the subdistribution of a competing risk. J Am Stat Assoc. 1997;94(446):496–509. , .
- Severe sepsis in managed care: analysis of incidence, one‐year mortality, and associated costs of care. J Manag Care Pharm. 2004;10(6):521–530. , , .
- Detailed cost analysis of care for survivors of severe sepsis. Crit Care Med. 2004;32(4):981–985. , , , , , .
- Readmissions among patients with severe sepsis/septic shock among inner‐city minority New Yorkers. Chest. 2012;142:286A. , , , , et al.
- Readmission and late mortality after pediatric severe sepsis. Pediatrics. 2009;123(3):849–857. , , .
- Rehospitalizations among patients in the Medicare fee‐for‐service program. N Engl J Med. 2009;360(14):1418–1428. , , .
- Post‐hospital syndrome—an acquired, transient condition of generalized risk. N Engl J Med. 2013;368(2):100–102. .
- Definitions for sepsis and organ failure and guidelines for the use of innovative therapies in sepsis. The ACCP/SCCM Consensus Conference Committee. American College of Chest Physicians/Society of Critical Care Medicine. Chest. 1992;101(6):1644–1655. , , , et al.
- The effect of age on the development and outcome of adult sepsis. Crit Care Med. 2006;34(1):15–21. , , .
- Multiple systems organ failure: failure of host defense homeostasis. Crit Care Clin. 1989;5(2):199–220. , .
- Pathophysiology of sepsis. Am J Pathol. 2007;170(5):1435–1444. .
- Surviving intensive care: a report from the 2002 Brussels Roundtable. Intensive Care Med. 2003;29(3):368–377. , .
- The encephalopathy in sepsis. Crit Care Clin. 2008;24(1):67–82, viii. , , .
- Sepsis‐associated encephalopathy. Nat Rev Neurol. 2012;8(10):557–566. , .
- Improving long‐term outcomes after discharge from intensive care unit: report from a stakeholders' conference. Crit Care Med. 2012;40(2):502–509. , , , et al.
- The association between sepsis and potential medical injury among hospitalized patients. Chest. 2012;142(3):606–613. , , , , .
- Three‐year outcomes for Medicare beneficiaries who survive intensive care. JAMA. 2010;303(9):849–856. , , , , , .
- Does acute organ dysfunction predict patient‐centered outcomes? Chest. 2002;121(6):1963–1971. , , , , , .
- Spurious inferences about long‐term outcomes: the case of severe sepsis and geriatric conditions. Am J Respir Crit Care Med. 2012;185(8):835–841. , , , .
- Population trends in the incidence and outcomes of acute myocardial infarction. N Engl J Med. 2010;362(23):2155–2165. , , , , , .
- Outpatient electronic health records and the clinical care and outcomes of patients with diabetes mellitus. Ann Intern Med. 2012;157(7):482–489. , , , et al.,
- Similarity of the adult Kaiser Permanente membership in Northern California to the insured and general population in Northern California: statistics from the 2009 California Health Interview Survey. Internal Division of Research Report. Oakland, CA: Kaiser Permanente Division of Research; January 24, 2012. Available at: http://www.dor.kaiser.org/external/chis_non_kp_2009. Accessed January 20, 2013. .
- Association of diagnostic coding with trends in hospitalizations and mortality of patients with pneumonia, 2003–2009. JAMA. 2012;307(13):1405–1413. , , , , .
- Finding pure and simple truths with administrative data. JAMA. 2012;307(13):1433–1435. , .
- Cost savings attributable to reductions in intensive care unit length of stay for mechanically ventilated patients. Med Care. 2008;46(12):1226–1233. , , , .
© 2014 Society of Hospital Medicine
Electronic Order Set for AMI
Although the prevalence of coronary heart disease and death from acute myocardial infarction (AMI) have declined steadily, about 935,000 heart attacks still occur annually in the United States, with approximately one‐third of these being fatal.[1, 2, 3] Studies have demonstrated decreased 30‐day and longer‐term mortality in AMI patients who receive evidence‐based treatment, including aspirin, ‐blockers, angiotensin‐converting enzyme (ACE) inhibitors or angiotensin receptor blockers (ARBs), anticoagulation therapy, and statins.[4, 5, 6, 7] Despite clinical practice guidelines (CPGs) outlining evidence‐based care and considerable efforts to implement processes that improve patient outcomes, delivery of effective therapy remains suboptimal.[8] For example, the Hospital Quality Alliance Program[9] found that in AMI patients, use of aspirin on admission was only 81% to 92%, ‐blocker on admission 75% to 85%, and ACE inhibitors for left ventricular dysfunction 71% to 74%.
Efforts to increase adherence to CPGs and improve patient outcomes in AMI have resulted in variable degrees of success. They include promotion of CPGs,[4, 5, 6, 7] physician education with feedback, report cards, care paths, registries,[10] Joint Commission standardized measures,[11] and paper checklists or order sets (OS).[12, 13]
In this report, we describe the association between use of an evidence‐based, electronic OS for AMI (AMI‐OS) and better adherence to CPGs. This AMI‐OS was implemented in the inpatient electronic medical records (EMRs) of a large integrated healthcare delivery system, Kaiser Permanente Northern California (KPNC). The purpose of our investigation was to determine (1) whether use of the AMI‐OS was associated with improved AMI processes and patient outcomes, and (2) whether these associations persisted after risk adjustment using a comprehensive severity of illness scoring system.
MATERIALS AND METHODS
This project was approved by the KPNC institutional review board.
Under a mutual exclusivity arrangement, salaried physicians of The Permanente Medical Group, Inc., care for 3.4 million Kaiser Foundation Health Plan, Inc. members at facilities owned by Kaiser Foundation Hospitals, Inc. All KPNC facilities employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere.[14] Our setting consisted of 21 KPNC hospitals described in previous reports,[15, 16, 17, 18] using the same commercially available EMR system that includes computerized physician order entry (CPOE). Deployment of the customized inpatient Epic EMR (
In this EMR's CPOE, physicians have options to select individual orders (a la carte) or they can utilize an OS, which is a collection of the most appropriate orders associated with specific diagnoses, procedures, or treatments. The evidence‐based AMI‐OS studied in this project was developed by a multidisciplinary team (for detailed components see Supporting Appendix 1Appendix 5 in the online version of this article).
Our study focused on the first set of hospital admission orders for patients with AMI. The study sample consisted of patients meeting these criteria: (1) age 18 years at admission; (2) admitted to a KPNC hospital for an overnight stay between September 28, 2008 and December 31, 2010; (3) principal diagnosis was AMI (International Classification of Diseases, 9th Revision [ICD‐9][19] codes 410.00, 01, 10, 11, 20, 21, 30, 31, 40, 41, 50, 51, 60, 61, 70, 71, 80, 90, and 91); and (4) KPHC had been operational at the hospital for at least 3 months to be included (for assembly descriptions see Supporting Appendices 15 in the online version of this article). At the study hospitals, troponin I was measured using the Beckman Access AccuTnI assay (Beckman Coulter, Inc., Brea, CA), whose upper reference limit (99th percentile) is 0.04 ng/mL. We excluded patients initially hospitalized for AMI at a non‐KPNC site and transferred into a study hospital.
The data processing methods we employed have been detailed elsewhere.[14, 15, 17, 20, 21, 22] The dependent outcome variables were total hospital length of stay, inpatient mortality, 30‐day mortality, and all‐cause rehospitalization within 30 days of discharge. Linked state mortality data were unavailable for the entire study period, so we ascertained 30‐day mortality based on the combination of KPNC patient demographic data and publicly available Social Security Administration decedent files. We ascertained rehospitalization by scanning KPNC hospitalization databases, which also track out‐of‐plan use.
The dependent process variables were use of aspirin within 24 hours of admission, ‐blockers, anticoagulation, ACE inhibitors or ARBs, and statins. The primary independent variable of interest was whether or not the admitting physician employed the AMI‐OS when admission orders were entered. Consequently, this variable is dichotomous (AMI‐OS vs a la carte).
We controlled for acute illness severity and chronic illness burden using a recent modification[22] of an externally validated risk‐adjustment system applicable to all hospitalized patients.[15, 16, 23, 24, 25] Our methodology included vital signs, neurological status checks, and laboratory test results obtained in the 72 hours preceding hospital admission; comorbidities were captured longitudinally using data from the year preceding hospitalization (for comparison purposes, we also assigned a Charlson Comorbidity Index score[26]).
End‐of‐life care directives are mandatory on admission at KPNC hospitals. Physicians have 4 options: full code, partial code, do not resuscitate, and comfort care only. Because of small numbers in some categories, we collapsed these 4 categories into full code and not full code. Because patients' care directives may change, we elected to capture the care directive in effect when a patient first entered a hospital unit other than the emergency department (ED).
Two authors (M.B., P.C.L.), one of whom is a board‐certified cardiologist, reviewed all admission electrocardiograms and made a consensus determination as to whether or not criteria for ST‐segment elevation myocardial infarction (STEMI) were present (ie, new ST‐segment elevation or left bundle branch block); we also reviewed the records of all patients with missing troponin I data to confirm the AMI diagnosis.
Statistical Methods
We performed unadjusted comparisons between AMI‐OS and nonAMI‐OS patients using the t test or the [2] test, as appropriate.
We hypothesized that the AMI‐OS plays a mediating role on patient outcomes through its effect on adherence to recommended treatment. We evaluated this hypothesis for inpatient mortality by first fitting a multivariable logistic regression model for inpatient mortality as the outcome and either the 5 evidence‐based therapies or the total number of evidence‐based therapies used (ranging from 02, 3, 4, or 5) as the dependent variable controlling for age, gender, presence of STEMI, troponin I, comorbidities, illness severity, ED length of stay (LOS), care directive status, and timing of cardiac catheterization referral as covariates to confirm the protective effect of these therapies on mortality. We then used the same model to estimate the effect of AMI‐OS on inpatient mortality, substituting the therapies with AMI‐OS as the dependent variable and using the same covariates. Last, we included both the therapies and the AMI‐OS in the model to evaluate their combined effects.[27]
We used 2 different methods to estimate the effects of AMI‐OS and number of therapies provided on the outcomes while adjusting for observed baseline differences between the 2 groups of patients: propensity risk score matching, which estimates the average treatment effect for the treated,[28, 29] and inverse probability of treatment weighting, which is used to estimate the average treatment effect.[30, 31, 32] The propensity score was defined as the probability of receiving the intervention for a patient with specific predictive factors.[33, 34] We computed a propensity score for each patient by using logistic regression, with the dependent variable being receipt of AMI‐OS and the independent variables being the covariates used for the multivariate logistic regression as well as ICD‐9 code for final diagnosis. We calculated the Mahalanobis distance between patients who received AMI‐OS (cases) and patients who did not received AMI‐OS (controls) using the same set of covariates. We matched each case to a single control within the same facility based on the nearest available Mahalanobis metric matching within calipers defied as the maximum width of 0.2 standard deviations of the logit of the estimated propensity score.[29, 35] We estimated the odds ratios for the binary dependent variables based on a conditional logistic regression model to account for the matched pairs design.[28] We used a generalized linear model with the log‐transformed LOS as the outcome to estimate the ratio of the LOS geometric mean of the cases to the controls. We calculated the relative risk for patients receiving AMI‐OS via the inverse probability weighting method by first defining a weight for each patient. [We assigned a weight of 1/psi to patients who received the AMI‐OS and a weight of 1/(1psi) to patients who did not receive the AMI‐OS, where psi denotes the propensity score for patient i]. We used a logistic regression model for the binary dependent variables with the same set of covariates described above to estimate the adjusted odds ratios while weighting each observation by its corresponding weight. Last, we used a weighted generalized linear model to estimate the AMI‐OS effect on the log‐transformed LOS.
RESULTS
Table 1 summarizes the characteristics of the 5879 patients. It shows that AMI‐OS patients were more likely to receive evidence‐based therapies for AMI (aspirin, ‐blockers, ACE inhibitors or ARBs, anticoagulation, and statins) and had a 46% lower mortality rate in hospital (3.51 % vs 6.52%) and 33% lower rate at 30 days (5.66% vs 8.48%). AMI‐OS patients were also found to be at lower risk for an adverse outcome than nonAMI‐OS patients. The AMI‐OS patients had lower peak troponin I values, severity of illness (lower Laboratory‐Based Acute Physiology Score, version 2 [LAPS2] scores), comorbidity burdens (lower Comorbidity Point Score, version 2 [COPS2] and Charlson scores), and global predicted mortality risk. AMI‐OS patients were also less likely to have required intensive care. AMI‐OS patients were at higher risk of death than nonAMI‐OS patients with respect to only 1 variable (being full code at the time of admission), but although this difference was statistically significant, it was of minor clinical impact (86% vs 88%).
Patients Initially Managed Using | P Valuea | ||
---|---|---|---|
AMI Order Set, N=3,531b | A La Carte Orders, N=2,348b | ||
| |||
Age, y, median (meanSD) | 70 (69.413.8) | 70 (69.213.8) | 0.5603 |
Age (% >65 years) | 2,134 (60.4%) | 1,415 (60.3%) | 0.8949 |
Sex (% male) | 2,202 (62.4%) | 1,451 (61.8%) | 0.6620 |
STEMI (% with)c | 166 (4.7%) | 369 (15.7%) | <0.0001 |
Troponin I (% missing) | 111 (3.1%) | 151 (6.4%) | <0.0001 |
Troponin I median (meanSD) | 0.57 (3.08.2) | 0.27 (2.58.9) | 0.0651 |
Charlson score median (meanSD)d | 2.0 (2.51.5) | 2.0 (2.71.6) | <0.0001 |
COPS2, median (meanSD)e | 14.0 (29.831.7) | 17.0 (34.334.4) | <0.0001 |
LAPS2, median (meanSD)e | 0.0 (35.643.5) | 27.0 (40.948.1) | <0.0001 |
Length of stay in ED, h, median (meanSD) | 5.7 (5.93.0) | 5.7 (5.43.1) | <0.0001 |
Patients receiving aspirin within 24 hoursf | 3,470 (98.3%) | 2,202 (93.8%) | <0.0001 |
Patients receiving anticoagulation therapyf | 2,886 (81.7%) | 1,846 (78.6%) | 0.0032 |
Patients receiving ‐blockersf | 3,196 (90.5%) | 1,926 (82.0%) | <0.0001 |
Patients receiving ACE inhibitors or ARBsf | 2,395 (67.8%) | 1,244 (53.0%) | <0.0001 |
Patients receiving statinsf | 3,337 (94.5%) | 1,975 (84.1%) | <0.0001 |
Patient received 1 or more therapies | 3,531 (100.0%) | 2,330 (99.2%) | <0.0001 |
Patient received 2 or more therapies | 3,521 (99.7%) | 2,266 (96.5%) | <0.0001 |
Patient received 3 or more therapies | 3,440 (97.4%) | 2,085 (88.8%) | <0.0001 |
Patient received 4 or more therapies | 3,015 (85.4%) | 1,646 (70.1%) | <0.0001 |
Patient received all 5 therapies | 1,777 (50.3%) | 866 (35.9%) | <0.0001 |
Predicted mortality risk, %, median, (meanSD)f | 0.86 (3.27.4) | 1.19 (4.810.8) | <0.0001 |
Full code at time of hospital entry (%)g | 3,041 (86.1%) | 2,066 (88.0%) | 0.0379 |
Admitted to ICU (%)i | |||
Direct admit | 826 (23.4%) | 567 (24.2%) | 0.5047 |
Unplanned transfer | 222 (6.3%) | 133 (5.7%) | 0.3262 |
Ever | 1,283 (36.3%) | 1,169 (49.8%) | <0.0001 |
Length of stay, h, median (meanSD) | 68.3 (109.4140.9) | 68.9 (113.8154.3) | 0.2615 |
Inpatient mortality (%) | 124 (3.5%) | 153 (6.5%) | <0.0001 |
30‐day mortality (%) | 200 (5.7%) | 199 (8.5%) | <0.0001 |
All‐cause rehospitalization within 30 days (%) | 576 (16.3%) | 401 (17.1%) | 0.4398 |
Cardiac catheterization procedure referral timing | |||
1 day preadmission to discharge | 2,018 (57.2%) | 1,348 (57.4%) | 0.1638 |
2 days preadmission or earlier | 97 (2.8%) | 87 (3.7%) | |
After discharge | 149 (4.2%) | 104 (4.4%) | |
No referral | 1,267 (35.9%) | 809 (34.5%) |
Table 2 shows the result of a logistic regression model in which the dependent variable was inpatient mortality and either the 5 evidence‐based therapies or the total number of evidence‐based therapies are the dependent variables. ‐blocker, statin, and ACE inhibitor or ARB therapies all had a protective effect on mortality, with odds ratios ranging from 0.48 (95% confidence interval [CI]: 0.36‐0.64), 0.63 (95% CI: 0.45‐0.89), and 0.40 (95% CI: 0.30‐0.53), respectively. An increased number of therapies also had a beneficial effect on inpatient mortality, with patients having 3 or more of the evidence‐based therapies showing an adjusted odds ratio (AOR) of 0.49 (95% CI: 0.33‐0.73), 4 or more therapies an AOR of 0.29 (95% CI: 0.20‐0.42), and 0.17 (95% CI: 0.11‐0.25) for 5 or more therapies.
Multiple Therapies Effect | Individual Therapies Effect | |||
---|---|---|---|---|
Outcome | Death | Death | ||
Number of outcomes | 277 | 277 | ||
AORa | 95% CIb | AORa | 95% CIb | |
| ||||
Age in years | ||||
1839 | Ref | Ref | ||
4064 | 1.02 | (0.147.73) | 1.01 | (0.137.66) |
6584 | 4.05 | (0.5529.72) | 3.89 | (0.5328.66) |
85+ | 4.99 | (0.6737.13) | 4.80 | (0.6435.84) |
Sex | ||||
Female | Ref | |||
Male | 1.05 | (0.811.37) | 1.07 | (0.821.39) |
STEMIc | ||||
Absent | Ref | Ref | ||
Present | 4.00 | (2.755.81) | 3.86 | (2.645.63) |
Troponin I | ||||
0.1 ng/ml | Ref | Ref | ||
>0.1 ng/ml | 1.01 | (0.721.42) | 1.02 | (0.731.43) |
COPS2d (AOR per 10 points) | 1.05 | (1.011.08) | 1.04 | (1.011.08) |
LAPS2d (AOR per 10 points) | 1.09 | (1.061.11) | 1.09 | (1.061.11) |
ED LOSe (hours) | ||||
<6 | Ref | Ref | ||
67 | 0.74 | (0.531.03) | 0.76 | (0.541.06) |
>=12 | 0.82 | (0.391.74) | 0.83 | (0.391.78) |
Code Statusf | ||||
Full Code | Ref | |||
Not Full Code | 1.08 | (0.781.49) | 1.09 | (0.791.51) |
Cardiac procedure referral | ||||
None during stay | Ref | |||
1 day pre adm until discharge | 0.40 | (0.290.54) | 0.39 | (0.280.53) |
Number of therapies received | ||||
2 or less | Ref | |||
3 | 0.49 | (0.330.73) | ||
4 | 0.29 | (0.200.42) | ||
5 | 0.17 | (0.110.25) | ||
Aspirin therapy | 0.80 | (0.491.32) | ||
Anticoagulation therapy | 0.86 | (0.641.16) | ||
Beta Blocker therapy | 0.48 | (0.360.64) | ||
Statin therapy | 0.63 | (0.450.89) | ||
ACE inhibitors or ARBs | 0.40 | (0.300.53) | ||
C Statistic | 0.814 | 0.822 | ||
Hosmer‐Lemeshow p value | 0.509 | 0.934 |
Table 3 shows that the use of the AMI‐OS is protective, with an AOR of 0.59 and a 95% CI of 0.45‐0.76. Table 3 also shows that the most potent predictors were comorbidity burden (AOR: 1.07, 95% CI: 1.03‐1.10 per 10 COPS2 points), severity of illness (AOR: 1.09, 95% CI: 1.07‐1.12 per 10 LAPS2 points), STEMI (AOR: 3.86, 95% CI: 2.68‐5.58), and timing of cardiac catheterization referral occurring immediately prior to or during the admission (AOR: 0.37, 95% CI: 0.27‐0.51). The statistical significance of the AMI‐OS effect disappears when both AMI‐OS and the individual therapies are included in the same model (see Supporting Information, Appendices 15, in the online version of this article).
Outcome | Death | |
---|---|---|
Number of outcomes | 277 | |
AORa | 95% CIb | |
| ||
Age in years | ||
1839 | Ref | |
4064 | 1.16 | (0.158.78) |
6584 | 4.67 | (0.6334.46) |
85+ | 5.45 | (0.7340.86) |
Sex | ||
Female | Ref | |
Male | 1.05 | (0.811.36) |
STEMIc | ||
Absent | Ref | |
Present | 3.86 | (2.685.58) |
Troponin I | ||
0.1 ng/ml | Ref | |
>0.1 ng/ml | 1.16 | (0.831.62) |
COPS2d (AOR per 10 points) | 1.07 | (1.031.10) |
LAPS2d (AOR per 10 points) | 1.09 | (1.071.12) |
ED LOSe (hours) | ||
<6 | Ref | |
67 | 0.72 | (0.521.00) |
>=12 | 0.70 | (0.331.48) |
Code statusf | ||
Full code | Ref | |
Not full code | 1.22 | (0.891.68) |
Cardiac procedure referral | ||
None during stay | Ref | |
1 day pre adm until discharge | 0.37 | (0.270.51) |
Order set employedg | ||
No | Ref | |
Yes | 0.59 | (0.450.76) |
C Statistic | 0.792 | |
Hosmer‐Lemeshow p value | 0.273 |
Table 4 shows separately the average treatment effect (ATE) and average treatment effect for the treated (ATT) of AMI‐OS and of increasing number of therapies on other outcomes (30‐day mortality, LOS, and readmission). Both the ATE and ATT show that the use of the AMI‐OS was significantly protective with respect to mortality and total hospital LOS but not significant with respect to readmission. The effect of the number of therapies on mortality is significantly higher with increasing number of therapies. For example, patients who received 5 therapies had an average treatment effect on 30‐day inpatient mortality of 0.23 (95% CI: 0.15‐0.35) compared to 0.64 (95% CI: 0.43‐0.96) for 3 therapies, almost a 3‐fold difference. The effects of increasing number of therapies were not significant for LOS or readmission. A sensitivity analysis in which the 535 STEMI patients were removed showed essentially the same results, so it is not reported here.
Outcome | Order Seta | 3 Therapiesb | 4 Therapiesb | 5 Therapiesb |
---|---|---|---|---|
| ||||
Average treatment effectc | ||||
Inpatient mortality | 0.67 (0.520.86) | 0.64 (0.430.96) | 0.37 (0.250.54) | 0.23 (0.150.35) |
30‐day mortality | 0.77 (0.620.96) | 0.68 (0.480.98) | 0.34 (0.240.48) | 0.26 (0.180.37) |
Readmission | 1.03 (0.901.19) | 1.20 (0.871.66) | 1.19 (0.881.60) | 1.30 (0.961.76) |
LOS, ratio of the geometric means | 0.91 (0.870.95) | 1.16 (1.031.30) | 1.17 (1.051.30) | 1.12 (1.001.24) |
Average treatment effect on the treatedd | ||||
Inpatient mortality | 0.69 (0.520.92) | 0.35 (0.130.93) | 0.17 (0.070.43) | 0.08 (0.030.20) |
30‐day mortality | 0.84 (0.661.06) | 0.35 (0.150.79) | 0.17 (0.070.37) | 0.09 (0.040.20) |
Readmission | 1.02 (0.871.20) | 1.39 (0.852.26) | 1.36 (0.882.12) | 1.23 (0.801.89) |
LOS, ratio of the geometric meanse | 0.92 (0.870.97) | 1.18 (1.021.37) | 1.16 (1.011.33) | 1.04 (0.911.19) |
To further elucidate possible reasons why physicians did not use the AMI‐OS, the lead author reviewed 105 randomly selected records where the AMI‐OS was not used, 5 records from each of the 21 study hospitals. This review found that in 36% of patients, the AMI‐OS was not used because emergent catheterization or transfer to a facility with percutaneous coronary intervention capability occurred. Presence of other significant medical conditions, including critical illness, was the reason in 17% of these cases, patient or family refusal of treatments in 8%, issues around end‐of‐life care in 3%, and specific medical contraindications in 1%. In the remaining 34%, no reason for not using the AMI‐OS could be identified.
DISCUSSION
We evaluated the use of an evidence‐based electronic AMI‐OS embedded in a comprehensive EMR and found that it was beneficial. Its use was associated with increased adherence to evidence‐based therapies, which in turn were associated with improved outcomes. Using data from a large cohort of hospitalized AMI patients in 21 community hospitals, we were able to use risk adjustment that included physiologic illness severity to adjust for baseline mortality risk. Patients in whom the AMI‐OS was employed tended to be at lower risk; nonetheless, after controlling for confounding variables and adjusting for bias using propensity scores, the AMI‐OS was associated with increased use of evidence‐based therapies and decreased mortality. Most importantly, it appears that the benefits of the OS were not just due to increased receipt of individual recommended therapies, but to increased concurrent receipt of multiple recommended therapies.
Modern EMRs have great potential for significant improvements in the quality, efficiency, and safety of care provided,[36] and our study highlights this potential. However, a number of important limitations to our study must be considered. Although we had access to a very rich dataset, we could not control for all possible confounders, and our risk adjustment cannot match the level of information available to clinicians. In particular, the measurements available to us with respect to cardiac risk are limited. Thus, we have to recognize that the strength of our findings does not approximate that of a randomized trial, and one would expect that the magnitude of the beneficial association would fall under more controlled conditions. Resource limitations also did not permit us to gather more time course data (eg, sequential measurements of patient instability, cardiac damage, or use of recommended therapies), which could provide a better delineation of differences in both processes and outcomes.
Limitations also exist to the generalizability of the use of order sets in other settings that go beyond the availability of a comprehensive EMR. Our study population was cared for in a setting with an unusually high level of integration.[1] For example, KPNC has an elaborate administrative infrastructure for training in the use of the EMR as well as ensuring that order sets are not just evidence‐based, but that they are perceived by clinicians to be of significant value. This infrastructure, established to ensure physician buy‐in, may not be easy to replicate in smaller or less‐integrated settings. Thus, it is conceivable that factors other than the degree of support during the EMR deployments can affect rates of order set use.
Although our use of counterfactual methods included illness severity (LAPS2) and longitudinal comorbidity burden (COPS2), which are not yet available outside highly integrated delivery services employing comprehensive EMRs, it is possible they are insufficient. We cannot exclude the possibility that other biases or patient characteristics were present that led clinicians to preferentially employ the electronic order set in some patients but not in others. One could also argue that future studies should consider using overall adherence to recommended AMI treatment guidelines as a risk adjustment tool that would permit one to analyze what other factors may be playing a role in residual differences in patient outcomes. Last, one could object to our inclusion of STEMI patients; however, this was not a study on optimum treatment strategies for STEMI patients. Rather, it was a study on the impact on AMI outcomes of a specific component of computerized order entry outside the research setting.
Despite these limitations, we believe that our findings provide strong support for the continued use of electronic evidence‐based order sets in the inpatient medical setting. Once the initial implementation of a comprehensive EMR has occurred, deployment of these electronic order sets is a relatively inexpensive but effective method to foster compliance with evidence‐based care.
Future research in healthcare information technology can take a number of directions. One important area, of course, revolves around ways to promote enhanced physician adoption of EMRs. Our audit of records where the AMI‐OS was not used found that specific reasons for not using the order set (eg, treatment refusals, emergent intervention) were present in two‐thirds of the cases. This suggests that future analyses of adherence involving EMRs and CPOE implementation should take a more nuanced look at how order entry is actually enabled. It may be that understanding how order sets affect care enhances clinician acceptance and thus could serve as an incentive to EMR adoption. However, once an EMR is adopted, a need exists to continue evaluations such as this because, ultimately, the gold standard should be improved patient care processes and better outcomes for patients.
Acknowledgement
The authors give special thanks to Dr. Brian Hoberman for sponsoring this work, Dr. Alan S. Go for providing assistance with obtaining copies of electrocardiograms for review, Drs. Tracy Lieu and Vincent Liu for reviewing the manuscript, and Ms. Rachel Lesser for formatting the manuscript.
Disclosures: This work was supported by The Permanente Medical Group, Inc. and Kaiser Foundation Hospitals, Inc. The algorithms used to extract data and perform risk adjustment were developed with funding from the Sidney Garfield Memorial Fund (Early Detection of Impending Physiologic Deterioration in Hospitalized Patients, 1159518), the Agency for Healthcare Quality and Research (Rapid Clinical Snapshots From the EMR Among Pneumonia Patients, 1R01HS018480‐01), and the Gordon and Betty Moore Foundation (Early Detection of Impending Physiologic Deterioration: Electronic Early Warning System).
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Although the prevalence of coronary heart disease and death from acute myocardial infarction (AMI) have declined steadily, about 935,000 heart attacks still occur annually in the United States, with approximately one‐third of these being fatal.[1, 2, 3] Studies have demonstrated decreased 30‐day and longer‐term mortality in AMI patients who receive evidence‐based treatment, including aspirin, ‐blockers, angiotensin‐converting enzyme (ACE) inhibitors or angiotensin receptor blockers (ARBs), anticoagulation therapy, and statins.[4, 5, 6, 7] Despite clinical practice guidelines (CPGs) outlining evidence‐based care and considerable efforts to implement processes that improve patient outcomes, delivery of effective therapy remains suboptimal.[8] For example, the Hospital Quality Alliance Program[9] found that in AMI patients, use of aspirin on admission was only 81% to 92%, ‐blocker on admission 75% to 85%, and ACE inhibitors for left ventricular dysfunction 71% to 74%.
Efforts to increase adherence to CPGs and improve patient outcomes in AMI have resulted in variable degrees of success. They include promotion of CPGs,[4, 5, 6, 7] physician education with feedback, report cards, care paths, registries,[10] Joint Commission standardized measures,[11] and paper checklists or order sets (OS).[12, 13]
In this report, we describe the association between use of an evidence‐based, electronic OS for AMI (AMI‐OS) and better adherence to CPGs. This AMI‐OS was implemented in the inpatient electronic medical records (EMRs) of a large integrated healthcare delivery system, Kaiser Permanente Northern California (KPNC). The purpose of our investigation was to determine (1) whether use of the AMI‐OS was associated with improved AMI processes and patient outcomes, and (2) whether these associations persisted after risk adjustment using a comprehensive severity of illness scoring system.
MATERIALS AND METHODS
This project was approved by the KPNC institutional review board.
Under a mutual exclusivity arrangement, salaried physicians of The Permanente Medical Group, Inc., care for 3.4 million Kaiser Foundation Health Plan, Inc. members at facilities owned by Kaiser Foundation Hospitals, Inc. All KPNC facilities employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere.[14] Our setting consisted of 21 KPNC hospitals described in previous reports,[15, 16, 17, 18] using the same commercially available EMR system that includes computerized physician order entry (CPOE). Deployment of the customized inpatient Epic EMR (
In this EMR's CPOE, physicians have options to select individual orders (a la carte) or they can utilize an OS, which is a collection of the most appropriate orders associated with specific diagnoses, procedures, or treatments. The evidence‐based AMI‐OS studied in this project was developed by a multidisciplinary team (for detailed components see Supporting Appendix 1Appendix 5 in the online version of this article).
Our study focused on the first set of hospital admission orders for patients with AMI. The study sample consisted of patients meeting these criteria: (1) age 18 years at admission; (2) admitted to a KPNC hospital for an overnight stay between September 28, 2008 and December 31, 2010; (3) principal diagnosis was AMI (International Classification of Diseases, 9th Revision [ICD‐9][19] codes 410.00, 01, 10, 11, 20, 21, 30, 31, 40, 41, 50, 51, 60, 61, 70, 71, 80, 90, and 91); and (4) KPHC had been operational at the hospital for at least 3 months to be included (for assembly descriptions see Supporting Appendices 15 in the online version of this article). At the study hospitals, troponin I was measured using the Beckman Access AccuTnI assay (Beckman Coulter, Inc., Brea, CA), whose upper reference limit (99th percentile) is 0.04 ng/mL. We excluded patients initially hospitalized for AMI at a non‐KPNC site and transferred into a study hospital.
The data processing methods we employed have been detailed elsewhere.[14, 15, 17, 20, 21, 22] The dependent outcome variables were total hospital length of stay, inpatient mortality, 30‐day mortality, and all‐cause rehospitalization within 30 days of discharge. Linked state mortality data were unavailable for the entire study period, so we ascertained 30‐day mortality based on the combination of KPNC patient demographic data and publicly available Social Security Administration decedent files. We ascertained rehospitalization by scanning KPNC hospitalization databases, which also track out‐of‐plan use.
The dependent process variables were use of aspirin within 24 hours of admission, ‐blockers, anticoagulation, ACE inhibitors or ARBs, and statins. The primary independent variable of interest was whether or not the admitting physician employed the AMI‐OS when admission orders were entered. Consequently, this variable is dichotomous (AMI‐OS vs a la carte).
We controlled for acute illness severity and chronic illness burden using a recent modification[22] of an externally validated risk‐adjustment system applicable to all hospitalized patients.[15, 16, 23, 24, 25] Our methodology included vital signs, neurological status checks, and laboratory test results obtained in the 72 hours preceding hospital admission; comorbidities were captured longitudinally using data from the year preceding hospitalization (for comparison purposes, we also assigned a Charlson Comorbidity Index score[26]).
End‐of‐life care directives are mandatory on admission at KPNC hospitals. Physicians have 4 options: full code, partial code, do not resuscitate, and comfort care only. Because of small numbers in some categories, we collapsed these 4 categories into full code and not full code. Because patients' care directives may change, we elected to capture the care directive in effect when a patient first entered a hospital unit other than the emergency department (ED).
Two authors (M.B., P.C.L.), one of whom is a board‐certified cardiologist, reviewed all admission electrocardiograms and made a consensus determination as to whether or not criteria for ST‐segment elevation myocardial infarction (STEMI) were present (ie, new ST‐segment elevation or left bundle branch block); we also reviewed the records of all patients with missing troponin I data to confirm the AMI diagnosis.
Statistical Methods
We performed unadjusted comparisons between AMI‐OS and nonAMI‐OS patients using the t test or the [2] test, as appropriate.
We hypothesized that the AMI‐OS plays a mediating role on patient outcomes through its effect on adherence to recommended treatment. We evaluated this hypothesis for inpatient mortality by first fitting a multivariable logistic regression model for inpatient mortality as the outcome and either the 5 evidence‐based therapies or the total number of evidence‐based therapies used (ranging from 02, 3, 4, or 5) as the dependent variable controlling for age, gender, presence of STEMI, troponin I, comorbidities, illness severity, ED length of stay (LOS), care directive status, and timing of cardiac catheterization referral as covariates to confirm the protective effect of these therapies on mortality. We then used the same model to estimate the effect of AMI‐OS on inpatient mortality, substituting the therapies with AMI‐OS as the dependent variable and using the same covariates. Last, we included both the therapies and the AMI‐OS in the model to evaluate their combined effects.[27]
We used 2 different methods to estimate the effects of AMI‐OS and number of therapies provided on the outcomes while adjusting for observed baseline differences between the 2 groups of patients: propensity risk score matching, which estimates the average treatment effect for the treated,[28, 29] and inverse probability of treatment weighting, which is used to estimate the average treatment effect.[30, 31, 32] The propensity score was defined as the probability of receiving the intervention for a patient with specific predictive factors.[33, 34] We computed a propensity score for each patient by using logistic regression, with the dependent variable being receipt of AMI‐OS and the independent variables being the covariates used for the multivariate logistic regression as well as ICD‐9 code for final diagnosis. We calculated the Mahalanobis distance between patients who received AMI‐OS (cases) and patients who did not received AMI‐OS (controls) using the same set of covariates. We matched each case to a single control within the same facility based on the nearest available Mahalanobis metric matching within calipers defied as the maximum width of 0.2 standard deviations of the logit of the estimated propensity score.[29, 35] We estimated the odds ratios for the binary dependent variables based on a conditional logistic regression model to account for the matched pairs design.[28] We used a generalized linear model with the log‐transformed LOS as the outcome to estimate the ratio of the LOS geometric mean of the cases to the controls. We calculated the relative risk for patients receiving AMI‐OS via the inverse probability weighting method by first defining a weight for each patient. [We assigned a weight of 1/psi to patients who received the AMI‐OS and a weight of 1/(1psi) to patients who did not receive the AMI‐OS, where psi denotes the propensity score for patient i]. We used a logistic regression model for the binary dependent variables with the same set of covariates described above to estimate the adjusted odds ratios while weighting each observation by its corresponding weight. Last, we used a weighted generalized linear model to estimate the AMI‐OS effect on the log‐transformed LOS.
RESULTS
Table 1 summarizes the characteristics of the 5879 patients. It shows that AMI‐OS patients were more likely to receive evidence‐based therapies for AMI (aspirin, ‐blockers, ACE inhibitors or ARBs, anticoagulation, and statins) and had a 46% lower mortality rate in hospital (3.51 % vs 6.52%) and 33% lower rate at 30 days (5.66% vs 8.48%). AMI‐OS patients were also found to be at lower risk for an adverse outcome than nonAMI‐OS patients. The AMI‐OS patients had lower peak troponin I values, severity of illness (lower Laboratory‐Based Acute Physiology Score, version 2 [LAPS2] scores), comorbidity burdens (lower Comorbidity Point Score, version 2 [COPS2] and Charlson scores), and global predicted mortality risk. AMI‐OS patients were also less likely to have required intensive care. AMI‐OS patients were at higher risk of death than nonAMI‐OS patients with respect to only 1 variable (being full code at the time of admission), but although this difference was statistically significant, it was of minor clinical impact (86% vs 88%).
Patients Initially Managed Using | P Valuea | ||
---|---|---|---|
AMI Order Set, N=3,531b | A La Carte Orders, N=2,348b | ||
| |||
Age, y, median (meanSD) | 70 (69.413.8) | 70 (69.213.8) | 0.5603 |
Age (% >65 years) | 2,134 (60.4%) | 1,415 (60.3%) | 0.8949 |
Sex (% male) | 2,202 (62.4%) | 1,451 (61.8%) | 0.6620 |
STEMI (% with)c | 166 (4.7%) | 369 (15.7%) | <0.0001 |
Troponin I (% missing) | 111 (3.1%) | 151 (6.4%) | <0.0001 |
Troponin I median (meanSD) | 0.57 (3.08.2) | 0.27 (2.58.9) | 0.0651 |
Charlson score median (meanSD)d | 2.0 (2.51.5) | 2.0 (2.71.6) | <0.0001 |
COPS2, median (meanSD)e | 14.0 (29.831.7) | 17.0 (34.334.4) | <0.0001 |
LAPS2, median (meanSD)e | 0.0 (35.643.5) | 27.0 (40.948.1) | <0.0001 |
Length of stay in ED, h, median (meanSD) | 5.7 (5.93.0) | 5.7 (5.43.1) | <0.0001 |
Patients receiving aspirin within 24 hoursf | 3,470 (98.3%) | 2,202 (93.8%) | <0.0001 |
Patients receiving anticoagulation therapyf | 2,886 (81.7%) | 1,846 (78.6%) | 0.0032 |
Patients receiving ‐blockersf | 3,196 (90.5%) | 1,926 (82.0%) | <0.0001 |
Patients receiving ACE inhibitors or ARBsf | 2,395 (67.8%) | 1,244 (53.0%) | <0.0001 |
Patients receiving statinsf | 3,337 (94.5%) | 1,975 (84.1%) | <0.0001 |
Patient received 1 or more therapies | 3,531 (100.0%) | 2,330 (99.2%) | <0.0001 |
Patient received 2 or more therapies | 3,521 (99.7%) | 2,266 (96.5%) | <0.0001 |
Patient received 3 or more therapies | 3,440 (97.4%) | 2,085 (88.8%) | <0.0001 |
Patient received 4 or more therapies | 3,015 (85.4%) | 1,646 (70.1%) | <0.0001 |
Patient received all 5 therapies | 1,777 (50.3%) | 866 (35.9%) | <0.0001 |
Predicted mortality risk, %, median, (meanSD)f | 0.86 (3.27.4) | 1.19 (4.810.8) | <0.0001 |
Full code at time of hospital entry (%)g | 3,041 (86.1%) | 2,066 (88.0%) | 0.0379 |
Admitted to ICU (%)i | |||
Direct admit | 826 (23.4%) | 567 (24.2%) | 0.5047 |
Unplanned transfer | 222 (6.3%) | 133 (5.7%) | 0.3262 |
Ever | 1,283 (36.3%) | 1,169 (49.8%) | <0.0001 |
Length of stay, h, median (meanSD) | 68.3 (109.4140.9) | 68.9 (113.8154.3) | 0.2615 |
Inpatient mortality (%) | 124 (3.5%) | 153 (6.5%) | <0.0001 |
30‐day mortality (%) | 200 (5.7%) | 199 (8.5%) | <0.0001 |
All‐cause rehospitalization within 30 days (%) | 576 (16.3%) | 401 (17.1%) | 0.4398 |
Cardiac catheterization procedure referral timing | |||
1 day preadmission to discharge | 2,018 (57.2%) | 1,348 (57.4%) | 0.1638 |
2 days preadmission or earlier | 97 (2.8%) | 87 (3.7%) | |
After discharge | 149 (4.2%) | 104 (4.4%) | |
No referral | 1,267 (35.9%) | 809 (34.5%) |
Table 2 shows the result of a logistic regression model in which the dependent variable was inpatient mortality and either the 5 evidence‐based therapies or the total number of evidence‐based therapies are the dependent variables. ‐blocker, statin, and ACE inhibitor or ARB therapies all had a protective effect on mortality, with odds ratios ranging from 0.48 (95% confidence interval [CI]: 0.36‐0.64), 0.63 (95% CI: 0.45‐0.89), and 0.40 (95% CI: 0.30‐0.53), respectively. An increased number of therapies also had a beneficial effect on inpatient mortality, with patients having 3 or more of the evidence‐based therapies showing an adjusted odds ratio (AOR) of 0.49 (95% CI: 0.33‐0.73), 4 or more therapies an AOR of 0.29 (95% CI: 0.20‐0.42), and 0.17 (95% CI: 0.11‐0.25) for 5 or more therapies.
Multiple Therapies Effect | Individual Therapies Effect | |||
---|---|---|---|---|
Outcome | Death | Death | ||
Number of outcomes | 277 | 277 | ||
AORa | 95% CIb | AORa | 95% CIb | |
| ||||
Age in years | ||||
1839 | Ref | Ref | ||
4064 | 1.02 | (0.147.73) | 1.01 | (0.137.66) |
6584 | 4.05 | (0.5529.72) | 3.89 | (0.5328.66) |
85+ | 4.99 | (0.6737.13) | 4.80 | (0.6435.84) |
Sex | ||||
Female | Ref | |||
Male | 1.05 | (0.811.37) | 1.07 | (0.821.39) |
STEMIc | ||||
Absent | Ref | Ref | ||
Present | 4.00 | (2.755.81) | 3.86 | (2.645.63) |
Troponin I | ||||
0.1 ng/ml | Ref | Ref | ||
>0.1 ng/ml | 1.01 | (0.721.42) | 1.02 | (0.731.43) |
COPS2d (AOR per 10 points) | 1.05 | (1.011.08) | 1.04 | (1.011.08) |
LAPS2d (AOR per 10 points) | 1.09 | (1.061.11) | 1.09 | (1.061.11) |
ED LOSe (hours) | ||||
<6 | Ref | Ref | ||
67 | 0.74 | (0.531.03) | 0.76 | (0.541.06) |
>=12 | 0.82 | (0.391.74) | 0.83 | (0.391.78) |
Code Statusf | ||||
Full Code | Ref | |||
Not Full Code | 1.08 | (0.781.49) | 1.09 | (0.791.51) |
Cardiac procedure referral | ||||
None during stay | Ref | |||
1 day pre adm until discharge | 0.40 | (0.290.54) | 0.39 | (0.280.53) |
Number of therapies received | ||||
2 or less | Ref | |||
3 | 0.49 | (0.330.73) | ||
4 | 0.29 | (0.200.42) | ||
5 | 0.17 | (0.110.25) | ||
Aspirin therapy | 0.80 | (0.491.32) | ||
Anticoagulation therapy | 0.86 | (0.641.16) | ||
Beta Blocker therapy | 0.48 | (0.360.64) | ||
Statin therapy | 0.63 | (0.450.89) | ||
ACE inhibitors or ARBs | 0.40 | (0.300.53) | ||
C Statistic | 0.814 | 0.822 | ||
Hosmer‐Lemeshow p value | 0.509 | 0.934 |
Table 3 shows that the use of the AMI‐OS is protective, with an AOR of 0.59 and a 95% CI of 0.45‐0.76. Table 3 also shows that the most potent predictors were comorbidity burden (AOR: 1.07, 95% CI: 1.03‐1.10 per 10 COPS2 points), severity of illness (AOR: 1.09, 95% CI: 1.07‐1.12 per 10 LAPS2 points), STEMI (AOR: 3.86, 95% CI: 2.68‐5.58), and timing of cardiac catheterization referral occurring immediately prior to or during the admission (AOR: 0.37, 95% CI: 0.27‐0.51). The statistical significance of the AMI‐OS effect disappears when both AMI‐OS and the individual therapies are included in the same model (see Supporting Information, Appendices 15, in the online version of this article).
Outcome | Death | |
---|---|---|
Number of outcomes | 277 | |
AORa | 95% CIb | |
| ||
Age in years | ||
1839 | Ref | |
4064 | 1.16 | (0.158.78) |
6584 | 4.67 | (0.6334.46) |
85+ | 5.45 | (0.7340.86) |
Sex | ||
Female | Ref | |
Male | 1.05 | (0.811.36) |
STEMIc | ||
Absent | Ref | |
Present | 3.86 | (2.685.58) |
Troponin I | ||
0.1 ng/ml | Ref | |
>0.1 ng/ml | 1.16 | (0.831.62) |
COPS2d (AOR per 10 points) | 1.07 | (1.031.10) |
LAPS2d (AOR per 10 points) | 1.09 | (1.071.12) |
ED LOSe (hours) | ||
<6 | Ref | |
67 | 0.72 | (0.521.00) |
>=12 | 0.70 | (0.331.48) |
Code statusf | ||
Full code | Ref | |
Not full code | 1.22 | (0.891.68) |
Cardiac procedure referral | ||
None during stay | Ref | |
1 day pre adm until discharge | 0.37 | (0.270.51) |
Order set employedg | ||
No | Ref | |
Yes | 0.59 | (0.450.76) |
C Statistic | 0.792 | |
Hosmer‐Lemeshow p value | 0.273 |
Table 4 shows separately the average treatment effect (ATE) and average treatment effect for the treated (ATT) of AMI‐OS and of increasing number of therapies on other outcomes (30‐day mortality, LOS, and readmission). Both the ATE and ATT show that the use of the AMI‐OS was significantly protective with respect to mortality and total hospital LOS but not significant with respect to readmission. The effect of the number of therapies on mortality is significantly higher with increasing number of therapies. For example, patients who received 5 therapies had an average treatment effect on 30‐day inpatient mortality of 0.23 (95% CI: 0.15‐0.35) compared to 0.64 (95% CI: 0.43‐0.96) for 3 therapies, almost a 3‐fold difference. The effects of increasing number of therapies were not significant for LOS or readmission. A sensitivity analysis in which the 535 STEMI patients were removed showed essentially the same results, so it is not reported here.
Outcome | Order Seta | 3 Therapiesb | 4 Therapiesb | 5 Therapiesb |
---|---|---|---|---|
| ||||
Average treatment effectc | ||||
Inpatient mortality | 0.67 (0.520.86) | 0.64 (0.430.96) | 0.37 (0.250.54) | 0.23 (0.150.35) |
30‐day mortality | 0.77 (0.620.96) | 0.68 (0.480.98) | 0.34 (0.240.48) | 0.26 (0.180.37) |
Readmission | 1.03 (0.901.19) | 1.20 (0.871.66) | 1.19 (0.881.60) | 1.30 (0.961.76) |
LOS, ratio of the geometric means | 0.91 (0.870.95) | 1.16 (1.031.30) | 1.17 (1.051.30) | 1.12 (1.001.24) |
Average treatment effect on the treatedd | ||||
Inpatient mortality | 0.69 (0.520.92) | 0.35 (0.130.93) | 0.17 (0.070.43) | 0.08 (0.030.20) |
30‐day mortality | 0.84 (0.661.06) | 0.35 (0.150.79) | 0.17 (0.070.37) | 0.09 (0.040.20) |
Readmission | 1.02 (0.871.20) | 1.39 (0.852.26) | 1.36 (0.882.12) | 1.23 (0.801.89) |
LOS, ratio of the geometric meanse | 0.92 (0.870.97) | 1.18 (1.021.37) | 1.16 (1.011.33) | 1.04 (0.911.19) |
To further elucidate possible reasons why physicians did not use the AMI‐OS, the lead author reviewed 105 randomly selected records where the AMI‐OS was not used, 5 records from each of the 21 study hospitals. This review found that in 36% of patients, the AMI‐OS was not used because emergent catheterization or transfer to a facility with percutaneous coronary intervention capability occurred. Presence of other significant medical conditions, including critical illness, was the reason in 17% of these cases, patient or family refusal of treatments in 8%, issues around end‐of‐life care in 3%, and specific medical contraindications in 1%. In the remaining 34%, no reason for not using the AMI‐OS could be identified.
DISCUSSION
We evaluated the use of an evidence‐based electronic AMI‐OS embedded in a comprehensive EMR and found that it was beneficial. Its use was associated with increased adherence to evidence‐based therapies, which in turn were associated with improved outcomes. Using data from a large cohort of hospitalized AMI patients in 21 community hospitals, we were able to use risk adjustment that included physiologic illness severity to adjust for baseline mortality risk. Patients in whom the AMI‐OS was employed tended to be at lower risk; nonetheless, after controlling for confounding variables and adjusting for bias using propensity scores, the AMI‐OS was associated with increased use of evidence‐based therapies and decreased mortality. Most importantly, it appears that the benefits of the OS were not just due to increased receipt of individual recommended therapies, but to increased concurrent receipt of multiple recommended therapies.
Modern EMRs have great potential for significant improvements in the quality, efficiency, and safety of care provided,[36] and our study highlights this potential. However, a number of important limitations to our study must be considered. Although we had access to a very rich dataset, we could not control for all possible confounders, and our risk adjustment cannot match the level of information available to clinicians. In particular, the measurements available to us with respect to cardiac risk are limited. Thus, we have to recognize that the strength of our findings does not approximate that of a randomized trial, and one would expect that the magnitude of the beneficial association would fall under more controlled conditions. Resource limitations also did not permit us to gather more time course data (eg, sequential measurements of patient instability, cardiac damage, or use of recommended therapies), which could provide a better delineation of differences in both processes and outcomes.
Limitations also exist to the generalizability of the use of order sets in other settings that go beyond the availability of a comprehensive EMR. Our study population was cared for in a setting with an unusually high level of integration.[1] For example, KPNC has an elaborate administrative infrastructure for training in the use of the EMR as well as ensuring that order sets are not just evidence‐based, but that they are perceived by clinicians to be of significant value. This infrastructure, established to ensure physician buy‐in, may not be easy to replicate in smaller or less‐integrated settings. Thus, it is conceivable that factors other than the degree of support during the EMR deployments can affect rates of order set use.
Although our use of counterfactual methods included illness severity (LAPS2) and longitudinal comorbidity burden (COPS2), which are not yet available outside highly integrated delivery services employing comprehensive EMRs, it is possible they are insufficient. We cannot exclude the possibility that other biases or patient characteristics were present that led clinicians to preferentially employ the electronic order set in some patients but not in others. One could also argue that future studies should consider using overall adherence to recommended AMI treatment guidelines as a risk adjustment tool that would permit one to analyze what other factors may be playing a role in residual differences in patient outcomes. Last, one could object to our inclusion of STEMI patients; however, this was not a study on optimum treatment strategies for STEMI patients. Rather, it was a study on the impact on AMI outcomes of a specific component of computerized order entry outside the research setting.
Despite these limitations, we believe that our findings provide strong support for the continued use of electronic evidence‐based order sets in the inpatient medical setting. Once the initial implementation of a comprehensive EMR has occurred, deployment of these electronic order sets is a relatively inexpensive but effective method to foster compliance with evidence‐based care.
Future research in healthcare information technology can take a number of directions. One important area, of course, revolves around ways to promote enhanced physician adoption of EMRs. Our audit of records where the AMI‐OS was not used found that specific reasons for not using the order set (eg, treatment refusals, emergent intervention) were present in two‐thirds of the cases. This suggests that future analyses of adherence involving EMRs and CPOE implementation should take a more nuanced look at how order entry is actually enabled. It may be that understanding how order sets affect care enhances clinician acceptance and thus could serve as an incentive to EMR adoption. However, once an EMR is adopted, a need exists to continue evaluations such as this because, ultimately, the gold standard should be improved patient care processes and better outcomes for patients.
Acknowledgement
The authors give special thanks to Dr. Brian Hoberman for sponsoring this work, Dr. Alan S. Go for providing assistance with obtaining copies of electrocardiograms for review, Drs. Tracy Lieu and Vincent Liu for reviewing the manuscript, and Ms. Rachel Lesser for formatting the manuscript.
Disclosures: This work was supported by The Permanente Medical Group, Inc. and Kaiser Foundation Hospitals, Inc. The algorithms used to extract data and perform risk adjustment were developed with funding from the Sidney Garfield Memorial Fund (Early Detection of Impending Physiologic Deterioration in Hospitalized Patients, 1159518), the Agency for Healthcare Quality and Research (Rapid Clinical Snapshots From the EMR Among Pneumonia Patients, 1R01HS018480‐01), and the Gordon and Betty Moore Foundation (Early Detection of Impending Physiologic Deterioration: Electronic Early Warning System).
Although the prevalence of coronary heart disease and death from acute myocardial infarction (AMI) have declined steadily, about 935,000 heart attacks still occur annually in the United States, with approximately one‐third of these being fatal.[1, 2, 3] Studies have demonstrated decreased 30‐day and longer‐term mortality in AMI patients who receive evidence‐based treatment, including aspirin, ‐blockers, angiotensin‐converting enzyme (ACE) inhibitors or angiotensin receptor blockers (ARBs), anticoagulation therapy, and statins.[4, 5, 6, 7] Despite clinical practice guidelines (CPGs) outlining evidence‐based care and considerable efforts to implement processes that improve patient outcomes, delivery of effective therapy remains suboptimal.[8] For example, the Hospital Quality Alliance Program[9] found that in AMI patients, use of aspirin on admission was only 81% to 92%, ‐blocker on admission 75% to 85%, and ACE inhibitors for left ventricular dysfunction 71% to 74%.
Efforts to increase adherence to CPGs and improve patient outcomes in AMI have resulted in variable degrees of success. They include promotion of CPGs,[4, 5, 6, 7] physician education with feedback, report cards, care paths, registries,[10] Joint Commission standardized measures,[11] and paper checklists or order sets (OS).[12, 13]
In this report, we describe the association between use of an evidence‐based, electronic OS for AMI (AMI‐OS) and better adherence to CPGs. This AMI‐OS was implemented in the inpatient electronic medical records (EMRs) of a large integrated healthcare delivery system, Kaiser Permanente Northern California (KPNC). The purpose of our investigation was to determine (1) whether use of the AMI‐OS was associated with improved AMI processes and patient outcomes, and (2) whether these associations persisted after risk adjustment using a comprehensive severity of illness scoring system.
MATERIALS AND METHODS
This project was approved by the KPNC institutional review board.
Under a mutual exclusivity arrangement, salaried physicians of The Permanente Medical Group, Inc., care for 3.4 million Kaiser Foundation Health Plan, Inc. members at facilities owned by Kaiser Foundation Hospitals, Inc. All KPNC facilities employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere.[14] Our setting consisted of 21 KPNC hospitals described in previous reports,[15, 16, 17, 18] using the same commercially available EMR system that includes computerized physician order entry (CPOE). Deployment of the customized inpatient Epic EMR (
In this EMR's CPOE, physicians have options to select individual orders (a la carte) or they can utilize an OS, which is a collection of the most appropriate orders associated with specific diagnoses, procedures, or treatments. The evidence‐based AMI‐OS studied in this project was developed by a multidisciplinary team (for detailed components see Supporting Appendix 1Appendix 5 in the online version of this article).
Our study focused on the first set of hospital admission orders for patients with AMI. The study sample consisted of patients meeting these criteria: (1) age 18 years at admission; (2) admitted to a KPNC hospital for an overnight stay between September 28, 2008 and December 31, 2010; (3) principal diagnosis was AMI (International Classification of Diseases, 9th Revision [ICD‐9][19] codes 410.00, 01, 10, 11, 20, 21, 30, 31, 40, 41, 50, 51, 60, 61, 70, 71, 80, 90, and 91); and (4) KPHC had been operational at the hospital for at least 3 months to be included (for assembly descriptions see Supporting Appendices 15 in the online version of this article). At the study hospitals, troponin I was measured using the Beckman Access AccuTnI assay (Beckman Coulter, Inc., Brea, CA), whose upper reference limit (99th percentile) is 0.04 ng/mL. We excluded patients initially hospitalized for AMI at a non‐KPNC site and transferred into a study hospital.
The data processing methods we employed have been detailed elsewhere.[14, 15, 17, 20, 21, 22] The dependent outcome variables were total hospital length of stay, inpatient mortality, 30‐day mortality, and all‐cause rehospitalization within 30 days of discharge. Linked state mortality data were unavailable for the entire study period, so we ascertained 30‐day mortality based on the combination of KPNC patient demographic data and publicly available Social Security Administration decedent files. We ascertained rehospitalization by scanning KPNC hospitalization databases, which also track out‐of‐plan use.
The dependent process variables were use of aspirin within 24 hours of admission, ‐blockers, anticoagulation, ACE inhibitors or ARBs, and statins. The primary independent variable of interest was whether or not the admitting physician employed the AMI‐OS when admission orders were entered. Consequently, this variable is dichotomous (AMI‐OS vs a la carte).
We controlled for acute illness severity and chronic illness burden using a recent modification[22] of an externally validated risk‐adjustment system applicable to all hospitalized patients.[15, 16, 23, 24, 25] Our methodology included vital signs, neurological status checks, and laboratory test results obtained in the 72 hours preceding hospital admission; comorbidities were captured longitudinally using data from the year preceding hospitalization (for comparison purposes, we also assigned a Charlson Comorbidity Index score[26]).
End‐of‐life care directives are mandatory on admission at KPNC hospitals. Physicians have 4 options: full code, partial code, do not resuscitate, and comfort care only. Because of small numbers in some categories, we collapsed these 4 categories into full code and not full code. Because patients' care directives may change, we elected to capture the care directive in effect when a patient first entered a hospital unit other than the emergency department (ED).
Two authors (M.B., P.C.L.), one of whom is a board‐certified cardiologist, reviewed all admission electrocardiograms and made a consensus determination as to whether or not criteria for ST‐segment elevation myocardial infarction (STEMI) were present (ie, new ST‐segment elevation or left bundle branch block); we also reviewed the records of all patients with missing troponin I data to confirm the AMI diagnosis.
Statistical Methods
We performed unadjusted comparisons between AMI‐OS and nonAMI‐OS patients using the t test or the [2] test, as appropriate.
We hypothesized that the AMI‐OS plays a mediating role on patient outcomes through its effect on adherence to recommended treatment. We evaluated this hypothesis for inpatient mortality by first fitting a multivariable logistic regression model for inpatient mortality as the outcome and either the 5 evidence‐based therapies or the total number of evidence‐based therapies used (ranging from 02, 3, 4, or 5) as the dependent variable controlling for age, gender, presence of STEMI, troponin I, comorbidities, illness severity, ED length of stay (LOS), care directive status, and timing of cardiac catheterization referral as covariates to confirm the protective effect of these therapies on mortality. We then used the same model to estimate the effect of AMI‐OS on inpatient mortality, substituting the therapies with AMI‐OS as the dependent variable and using the same covariates. Last, we included both the therapies and the AMI‐OS in the model to evaluate their combined effects.[27]
We used 2 different methods to estimate the effects of AMI‐OS and number of therapies provided on the outcomes while adjusting for observed baseline differences between the 2 groups of patients: propensity risk score matching, which estimates the average treatment effect for the treated,[28, 29] and inverse probability of treatment weighting, which is used to estimate the average treatment effect.[30, 31, 32] The propensity score was defined as the probability of receiving the intervention for a patient with specific predictive factors.[33, 34] We computed a propensity score for each patient by using logistic regression, with the dependent variable being receipt of AMI‐OS and the independent variables being the covariates used for the multivariate logistic regression as well as ICD‐9 code for final diagnosis. We calculated the Mahalanobis distance between patients who received AMI‐OS (cases) and patients who did not received AMI‐OS (controls) using the same set of covariates. We matched each case to a single control within the same facility based on the nearest available Mahalanobis metric matching within calipers defied as the maximum width of 0.2 standard deviations of the logit of the estimated propensity score.[29, 35] We estimated the odds ratios for the binary dependent variables based on a conditional logistic regression model to account for the matched pairs design.[28] We used a generalized linear model with the log‐transformed LOS as the outcome to estimate the ratio of the LOS geometric mean of the cases to the controls. We calculated the relative risk for patients receiving AMI‐OS via the inverse probability weighting method by first defining a weight for each patient. [We assigned a weight of 1/psi to patients who received the AMI‐OS and a weight of 1/(1psi) to patients who did not receive the AMI‐OS, where psi denotes the propensity score for patient i]. We used a logistic regression model for the binary dependent variables with the same set of covariates described above to estimate the adjusted odds ratios while weighting each observation by its corresponding weight. Last, we used a weighted generalized linear model to estimate the AMI‐OS effect on the log‐transformed LOS.
RESULTS
Table 1 summarizes the characteristics of the 5879 patients. It shows that AMI‐OS patients were more likely to receive evidence‐based therapies for AMI (aspirin, ‐blockers, ACE inhibitors or ARBs, anticoagulation, and statins) and had a 46% lower mortality rate in hospital (3.51 % vs 6.52%) and 33% lower rate at 30 days (5.66% vs 8.48%). AMI‐OS patients were also found to be at lower risk for an adverse outcome than nonAMI‐OS patients. The AMI‐OS patients had lower peak troponin I values, severity of illness (lower Laboratory‐Based Acute Physiology Score, version 2 [LAPS2] scores), comorbidity burdens (lower Comorbidity Point Score, version 2 [COPS2] and Charlson scores), and global predicted mortality risk. AMI‐OS patients were also less likely to have required intensive care. AMI‐OS patients were at higher risk of death than nonAMI‐OS patients with respect to only 1 variable (being full code at the time of admission), but although this difference was statistically significant, it was of minor clinical impact (86% vs 88%).
Patients Initially Managed Using | P Valuea | ||
---|---|---|---|
AMI Order Set, N=3,531b | A La Carte Orders, N=2,348b | ||
| |||
Age, y, median (meanSD) | 70 (69.413.8) | 70 (69.213.8) | 0.5603 |
Age (% >65 years) | 2,134 (60.4%) | 1,415 (60.3%) | 0.8949 |
Sex (% male) | 2,202 (62.4%) | 1,451 (61.8%) | 0.6620 |
STEMI (% with)c | 166 (4.7%) | 369 (15.7%) | <0.0001 |
Troponin I (% missing) | 111 (3.1%) | 151 (6.4%) | <0.0001 |
Troponin I median (meanSD) | 0.57 (3.08.2) | 0.27 (2.58.9) | 0.0651 |
Charlson score median (meanSD)d | 2.0 (2.51.5) | 2.0 (2.71.6) | <0.0001 |
COPS2, median (meanSD)e | 14.0 (29.831.7) | 17.0 (34.334.4) | <0.0001 |
LAPS2, median (meanSD)e | 0.0 (35.643.5) | 27.0 (40.948.1) | <0.0001 |
Length of stay in ED, h, median (meanSD) | 5.7 (5.93.0) | 5.7 (5.43.1) | <0.0001 |
Patients receiving aspirin within 24 hoursf | 3,470 (98.3%) | 2,202 (93.8%) | <0.0001 |
Patients receiving anticoagulation therapyf | 2,886 (81.7%) | 1,846 (78.6%) | 0.0032 |
Patients receiving ‐blockersf | 3,196 (90.5%) | 1,926 (82.0%) | <0.0001 |
Patients receiving ACE inhibitors or ARBsf | 2,395 (67.8%) | 1,244 (53.0%) | <0.0001 |
Patients receiving statinsf | 3,337 (94.5%) | 1,975 (84.1%) | <0.0001 |
Patient received 1 or more therapies | 3,531 (100.0%) | 2,330 (99.2%) | <0.0001 |
Patient received 2 or more therapies | 3,521 (99.7%) | 2,266 (96.5%) | <0.0001 |
Patient received 3 or more therapies | 3,440 (97.4%) | 2,085 (88.8%) | <0.0001 |
Patient received 4 or more therapies | 3,015 (85.4%) | 1,646 (70.1%) | <0.0001 |
Patient received all 5 therapies | 1,777 (50.3%) | 866 (35.9%) | <0.0001 |
Predicted mortality risk, %, median, (meanSD)f | 0.86 (3.27.4) | 1.19 (4.810.8) | <0.0001 |
Full code at time of hospital entry (%)g | 3,041 (86.1%) | 2,066 (88.0%) | 0.0379 |
Admitted to ICU (%)i | |||
Direct admit | 826 (23.4%) | 567 (24.2%) | 0.5047 |
Unplanned transfer | 222 (6.3%) | 133 (5.7%) | 0.3262 |
Ever | 1,283 (36.3%) | 1,169 (49.8%) | <0.0001 |
Length of stay, h, median (meanSD) | 68.3 (109.4140.9) | 68.9 (113.8154.3) | 0.2615 |
Inpatient mortality (%) | 124 (3.5%) | 153 (6.5%) | <0.0001 |
30‐day mortality (%) | 200 (5.7%) | 199 (8.5%) | <0.0001 |
All‐cause rehospitalization within 30 days (%) | 576 (16.3%) | 401 (17.1%) | 0.4398 |
Cardiac catheterization procedure referral timing | |||
1 day preadmission to discharge | 2,018 (57.2%) | 1,348 (57.4%) | 0.1638 |
2 days preadmission or earlier | 97 (2.8%) | 87 (3.7%) | |
After discharge | 149 (4.2%) | 104 (4.4%) | |
No referral | 1,267 (35.9%) | 809 (34.5%) |
Table 2 shows the result of a logistic regression model in which the dependent variable was inpatient mortality and either the 5 evidence‐based therapies or the total number of evidence‐based therapies are the dependent variables. ‐blocker, statin, and ACE inhibitor or ARB therapies all had a protective effect on mortality, with odds ratios ranging from 0.48 (95% confidence interval [CI]: 0.36‐0.64), 0.63 (95% CI: 0.45‐0.89), and 0.40 (95% CI: 0.30‐0.53), respectively. An increased number of therapies also had a beneficial effect on inpatient mortality, with patients having 3 or more of the evidence‐based therapies showing an adjusted odds ratio (AOR) of 0.49 (95% CI: 0.33‐0.73), 4 or more therapies an AOR of 0.29 (95% CI: 0.20‐0.42), and 0.17 (95% CI: 0.11‐0.25) for 5 or more therapies.
Multiple Therapies Effect | Individual Therapies Effect | |||
---|---|---|---|---|
Outcome | Death | Death | ||
Number of outcomes | 277 | 277 | ||
AORa | 95% CIb | AORa | 95% CIb | |
| ||||
Age in years | ||||
1839 | Ref | Ref | ||
4064 | 1.02 | (0.147.73) | 1.01 | (0.137.66) |
6584 | 4.05 | (0.5529.72) | 3.89 | (0.5328.66) |
85+ | 4.99 | (0.6737.13) | 4.80 | (0.6435.84) |
Sex | ||||
Female | Ref | |||
Male | 1.05 | (0.811.37) | 1.07 | (0.821.39) |
STEMIc | ||||
Absent | Ref | Ref | ||
Present | 4.00 | (2.755.81) | 3.86 | (2.645.63) |
Troponin I | ||||
0.1 ng/ml | Ref | Ref | ||
>0.1 ng/ml | 1.01 | (0.721.42) | 1.02 | (0.731.43) |
COPS2d (AOR per 10 points) | 1.05 | (1.011.08) | 1.04 | (1.011.08) |
LAPS2d (AOR per 10 points) | 1.09 | (1.061.11) | 1.09 | (1.061.11) |
ED LOSe (hours) | ||||
<6 | Ref | Ref | ||
67 | 0.74 | (0.531.03) | 0.76 | (0.541.06) |
>=12 | 0.82 | (0.391.74) | 0.83 | (0.391.78) |
Code Statusf | ||||
Full Code | Ref | |||
Not Full Code | 1.08 | (0.781.49) | 1.09 | (0.791.51) |
Cardiac procedure referral | ||||
None during stay | Ref | |||
1 day pre adm until discharge | 0.40 | (0.290.54) | 0.39 | (0.280.53) |
Number of therapies received | ||||
2 or less | Ref | |||
3 | 0.49 | (0.330.73) | ||
4 | 0.29 | (0.200.42) | ||
5 | 0.17 | (0.110.25) | ||
Aspirin therapy | 0.80 | (0.491.32) | ||
Anticoagulation therapy | 0.86 | (0.641.16) | ||
Beta Blocker therapy | 0.48 | (0.360.64) | ||
Statin therapy | 0.63 | (0.450.89) | ||
ACE inhibitors or ARBs | 0.40 | (0.300.53) | ||
C Statistic | 0.814 | 0.822 | ||
Hosmer‐Lemeshow p value | 0.509 | 0.934 |
Table 3 shows that the use of the AMI‐OS is protective, with an AOR of 0.59 and a 95% CI of 0.45‐0.76. Table 3 also shows that the most potent predictors were comorbidity burden (AOR: 1.07, 95% CI: 1.03‐1.10 per 10 COPS2 points), severity of illness (AOR: 1.09, 95% CI: 1.07‐1.12 per 10 LAPS2 points), STEMI (AOR: 3.86, 95% CI: 2.68‐5.58), and timing of cardiac catheterization referral occurring immediately prior to or during the admission (AOR: 0.37, 95% CI: 0.27‐0.51). The statistical significance of the AMI‐OS effect disappears when both AMI‐OS and the individual therapies are included in the same model (see Supporting Information, Appendices 15, in the online version of this article).
Outcome | Death | |
---|---|---|
Number of outcomes | 277 | |
AORa | 95% CIb | |
| ||
Age in years | ||
1839 | Ref | |
4064 | 1.16 | (0.158.78) |
6584 | 4.67 | (0.6334.46) |
85+ | 5.45 | (0.7340.86) |
Sex | ||
Female | Ref | |
Male | 1.05 | (0.811.36) |
STEMIc | ||
Absent | Ref | |
Present | 3.86 | (2.685.58) |
Troponin I | ||
0.1 ng/ml | Ref | |
>0.1 ng/ml | 1.16 | (0.831.62) |
COPS2d (AOR per 10 points) | 1.07 | (1.031.10) |
LAPS2d (AOR per 10 points) | 1.09 | (1.071.12) |
ED LOSe (hours) | ||
<6 | Ref | |
67 | 0.72 | (0.521.00) |
>=12 | 0.70 | (0.331.48) |
Code statusf | ||
Full code | Ref | |
Not full code | 1.22 | (0.891.68) |
Cardiac procedure referral | ||
None during stay | Ref | |
1 day pre adm until discharge | 0.37 | (0.270.51) |
Order set employedg | ||
No | Ref | |
Yes | 0.59 | (0.450.76) |
C Statistic | 0.792 | |
Hosmer‐Lemeshow p value | 0.273 |
Table 4 shows separately the average treatment effect (ATE) and average treatment effect for the treated (ATT) of AMI‐OS and of increasing number of therapies on other outcomes (30‐day mortality, LOS, and readmission). Both the ATE and ATT show that the use of the AMI‐OS was significantly protective with respect to mortality and total hospital LOS but not significant with respect to readmission. The effect of the number of therapies on mortality is significantly higher with increasing number of therapies. For example, patients who received 5 therapies had an average treatment effect on 30‐day inpatient mortality of 0.23 (95% CI: 0.15‐0.35) compared to 0.64 (95% CI: 0.43‐0.96) for 3 therapies, almost a 3‐fold difference. The effects of increasing number of therapies were not significant for LOS or readmission. A sensitivity analysis in which the 535 STEMI patients were removed showed essentially the same results, so it is not reported here.
Outcome | Order Seta | 3 Therapiesb | 4 Therapiesb | 5 Therapiesb |
---|---|---|---|---|
| ||||
Average treatment effectc | ||||
Inpatient mortality | 0.67 (0.520.86) | 0.64 (0.430.96) | 0.37 (0.250.54) | 0.23 (0.150.35) |
30‐day mortality | 0.77 (0.620.96) | 0.68 (0.480.98) | 0.34 (0.240.48) | 0.26 (0.180.37) |
Readmission | 1.03 (0.901.19) | 1.20 (0.871.66) | 1.19 (0.881.60) | 1.30 (0.961.76) |
LOS, ratio of the geometric means | 0.91 (0.870.95) | 1.16 (1.031.30) | 1.17 (1.051.30) | 1.12 (1.001.24) |
Average treatment effect on the treatedd | ||||
Inpatient mortality | 0.69 (0.520.92) | 0.35 (0.130.93) | 0.17 (0.070.43) | 0.08 (0.030.20) |
30‐day mortality | 0.84 (0.661.06) | 0.35 (0.150.79) | 0.17 (0.070.37) | 0.09 (0.040.20) |
Readmission | 1.02 (0.871.20) | 1.39 (0.852.26) | 1.36 (0.882.12) | 1.23 (0.801.89) |
LOS, ratio of the geometric meanse | 0.92 (0.870.97) | 1.18 (1.021.37) | 1.16 (1.011.33) | 1.04 (0.911.19) |
To further elucidate possible reasons why physicians did not use the AMI‐OS, the lead author reviewed 105 randomly selected records where the AMI‐OS was not used, 5 records from each of the 21 study hospitals. This review found that in 36% of patients, the AMI‐OS was not used because emergent catheterization or transfer to a facility with percutaneous coronary intervention capability occurred. Presence of other significant medical conditions, including critical illness, was the reason in 17% of these cases, patient or family refusal of treatments in 8%, issues around end‐of‐life care in 3%, and specific medical contraindications in 1%. In the remaining 34%, no reason for not using the AMI‐OS could be identified.
DISCUSSION
We evaluated the use of an evidence‐based electronic AMI‐OS embedded in a comprehensive EMR and found that it was beneficial. Its use was associated with increased adherence to evidence‐based therapies, which in turn were associated with improved outcomes. Using data from a large cohort of hospitalized AMI patients in 21 community hospitals, we were able to use risk adjustment that included physiologic illness severity to adjust for baseline mortality risk. Patients in whom the AMI‐OS was employed tended to be at lower risk; nonetheless, after controlling for confounding variables and adjusting for bias using propensity scores, the AMI‐OS was associated with increased use of evidence‐based therapies and decreased mortality. Most importantly, it appears that the benefits of the OS were not just due to increased receipt of individual recommended therapies, but to increased concurrent receipt of multiple recommended therapies.
Modern EMRs have great potential for significant improvements in the quality, efficiency, and safety of care provided,[36] and our study highlights this potential. However, a number of important limitations to our study must be considered. Although we had access to a very rich dataset, we could not control for all possible confounders, and our risk adjustment cannot match the level of information available to clinicians. In particular, the measurements available to us with respect to cardiac risk are limited. Thus, we have to recognize that the strength of our findings does not approximate that of a randomized trial, and one would expect that the magnitude of the beneficial association would fall under more controlled conditions. Resource limitations also did not permit us to gather more time course data (eg, sequential measurements of patient instability, cardiac damage, or use of recommended therapies), which could provide a better delineation of differences in both processes and outcomes.
Limitations also exist to the generalizability of the use of order sets in other settings that go beyond the availability of a comprehensive EMR. Our study population was cared for in a setting with an unusually high level of integration.[1] For example, KPNC has an elaborate administrative infrastructure for training in the use of the EMR as well as ensuring that order sets are not just evidence‐based, but that they are perceived by clinicians to be of significant value. This infrastructure, established to ensure physician buy‐in, may not be easy to replicate in smaller or less‐integrated settings. Thus, it is conceivable that factors other than the degree of support during the EMR deployments can affect rates of order set use.
Although our use of counterfactual methods included illness severity (LAPS2) and longitudinal comorbidity burden (COPS2), which are not yet available outside highly integrated delivery services employing comprehensive EMRs, it is possible they are insufficient. We cannot exclude the possibility that other biases or patient characteristics were present that led clinicians to preferentially employ the electronic order set in some patients but not in others. One could also argue that future studies should consider using overall adherence to recommended AMI treatment guidelines as a risk adjustment tool that would permit one to analyze what other factors may be playing a role in residual differences in patient outcomes. Last, one could object to our inclusion of STEMI patients; however, this was not a study on optimum treatment strategies for STEMI patients. Rather, it was a study on the impact on AMI outcomes of a specific component of computerized order entry outside the research setting.
Despite these limitations, we believe that our findings provide strong support for the continued use of electronic evidence‐based order sets in the inpatient medical setting. Once the initial implementation of a comprehensive EMR has occurred, deployment of these electronic order sets is a relatively inexpensive but effective method to foster compliance with evidence‐based care.
Future research in healthcare information technology can take a number of directions. One important area, of course, revolves around ways to promote enhanced physician adoption of EMRs. Our audit of records where the AMI‐OS was not used found that specific reasons for not using the order set (eg, treatment refusals, emergent intervention) were present in two‐thirds of the cases. This suggests that future analyses of adherence involving EMRs and CPOE implementation should take a more nuanced look at how order entry is actually enabled. It may be that understanding how order sets affect care enhances clinician acceptance and thus could serve as an incentive to EMR adoption. However, once an EMR is adopted, a need exists to continue evaluations such as this because, ultimately, the gold standard should be improved patient care processes and better outcomes for patients.
Acknowledgement
The authors give special thanks to Dr. Brian Hoberman for sponsoring this work, Dr. Alan S. Go for providing assistance with obtaining copies of electrocardiograms for review, Drs. Tracy Lieu and Vincent Liu for reviewing the manuscript, and Ms. Rachel Lesser for formatting the manuscript.
Disclosures: This work was supported by The Permanente Medical Group, Inc. and Kaiser Foundation Hospitals, Inc. The algorithms used to extract data and perform risk adjustment were developed with funding from the Sidney Garfield Memorial Fund (Early Detection of Impending Physiologic Deterioration in Hospitalized Patients, 1159518), the Agency for Healthcare Quality and Research (Rapid Clinical Snapshots From the EMR Among Pneumonia Patients, 1R01HS018480‐01), and the Gordon and Betty Moore Foundation (Early Detection of Impending Physiologic Deterioration: Electronic Early Warning System).
- Population trends in the incidence and outcomes of acute myocardial infarction. N Engl J Med. 2010;362(23):2155–2165. , , , , , .
- Twenty‐two‐year trends in incidence of myocardial infarction, coronary heart disease mortality, and case fatality in 4 US communities, 1987–2008. Circulation. 2012;125(15):1848–1857. , , , et al.
- Heart disease and stroke statistics—2012 update: a report from the American Heart Association. Circulation. 2012;125(1):e2–e220. , , , et al.
- ACC/AHA 2007 guidelines for the management of patients with unstable angina/non‐ST‐Elevation myocardial infarction: a report of the American College of Cardiology/American Heart Association Task Force on Practice Guidelines (Writing Committee to Revise the 2002 Guidelines for the Management of Patients With Unstable Angina/Non‐ST‐Elevation Myocardial Infarction) developed in collaboration with the American College of Emergency Physicians, the Society for Cardiovascular Angiography and Interventions, and the Society of Thoracic Surgeons endorsed by the American Association of Cardiovascular and Pulmonary Rehabilitation and the Society for Academic Emergency Medicine. J Am Coll Cardiol. 2007;50(7):e1–e157. , , , et al.
- 2007 focused update of the ACC/AHA 2004 guidelines for the management of patients with ST‐elevation myocardial infarction: a report of the American College of Cardiology/American Heart Association Task Force on Practice Guidelines. J Am Coll Cardiol. 2008;51(2):210–247. , , , et al.
- Association between adoption of evidence‐based treatment and survival for patients with ST‐elevation myocardial infarction. JAMA. 2011;305(16):1677–1684. , , , , , .
- Association of changes in clinical characteristics and management with improvement in survival among patients with ST‐elevation myocardial infarction. JAMA. 2012;308(10):998–1006. , , , et al.
- Changes in myocardial infarction guideline adherence as a function of patient risk: an end to paradoxical care? J Am Coll Cardiol. 2011;58(17):1760–1765. , , , et al.
- Care in U.S. hospitals—the Hospital Quality Alliance program. N Engl J Med. 2005;353(3):265–274. , , , .
- Challenges in the treatment of NSTEMI patients at high risk for both ischemic and bleeding events: insights from the ACTION Registry‐GWTG. J Am Coll Cardiol. 2011;57:E913–E913. , , et al.
- Quality of care in U.S. hospitals as reflected by standardized measures, 2002–2004. N Engl J Med. 2005;353(3):255–264. , , , , .
- Guideline‐based standardized care is associated with substantially lower mortality in medicare patients with acute myocardial infarction. J Am Coll Cardiol. 2005;46(7):1242–1248. , , .
- Impact of a standardized heart failure order set on mortality, readmission, and quality and costs of care. Int J Qual Health Care. 2010;22(6):437–444. , , , et al.
- Linking automated databases for research in managed care settings. Ann Intern Med. 1997;127(8 pt 2):719–724. .
- Risk adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , , , .
- Length of stay predictions: improvements through the use of automated laboratory and comorbidity variables. Med Care. 2010;48(8):739–744. , , , .
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6(2):74–80. , , , , , .
- Adverse outcomes associated with delayed intensive care unit transfers in an integrated healthcare system. J Hosp Med. 2012;7(3):224–230. , , , .
- International Classification of Diseases, 9th Revision‐Clinical Modification. 4th ed. 3 Vols. Los Angeles, CA: Practice Management Information Corporation; 2006.
- Anticoagulation therapy for stroke prevention in atrial fibrillation: how well do randomized trials translate into clinical practice? JAMA. 2003;290(20):2685–2692. , , , et al.
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated healthcare delivery system. Med Care. 2013;51(5):446–453. , , , , .
- Effect of choice of estimation method on inter‐hospital mortality rate comparisons. Med Care. 2010;48(5):456–485. , , .
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population. J Clin Epidemiol. 2010;63(7):798–803. , , , .
- Derivation and validation of a model to predict daily risk of death in hospital. Med Care. 2011;49(8):734–743. , , , , .
- Adapting a clinical comorbidity index for use with ICD‐9‐CM administrative databases. J Clin Epidemiol. 1992;45(6):613–619. , , .
- Introduction to Statistical Mediation Analysis. New York, NY: Lawrence Erlbaum Associates; 2008. .
- Nonparametric estimation of average treatment effects under exogenity: a review. Rev Econ Stat. 2004;86:25. .
- Design of Observational Studies. New York, NY: Springer Science+Business Media; 2010. .
- Balance diagnostics for comparing the distribution of baseline covariates between treatment groups in propensity‐score matched samples. Stat Med. 2009;28:24. .
- Estimation of regression coefficients when some regressors are not always observed. J Am Stat Assoc. 1994(89):846–866. , , .
- Stratification and weighting via the propensity score in estimation of causal treatment effects: a comparative study. Stat Med. 2004;23(19):2937–2960. , .
- Discussing hidden bias in observational studies. Ann Intern Med. 1991;115(11):901–905. .
- Propensity score methods for bias reduction in the comparison of a treatment to a non‐randomized control group. Stat Med. 1998;17(19):2265–2281. .
- A method/macro based on propensity score and Mahalanobis distance to reduce bias in treatment comparison in observational study, 2005. www.lexjansen.com/pharmasug/2006/publichealthresearch/pr05.pdf. Accessed on September 14, 2013. , , .
- Using health information technology to improve health care. Arch Intern Med. 2012;172(22):1728–1730. .
- Population trends in the incidence and outcomes of acute myocardial infarction. N Engl J Med. 2010;362(23):2155–2165. , , , , , .
- Twenty‐two‐year trends in incidence of myocardial infarction, coronary heart disease mortality, and case fatality in 4 US communities, 1987–2008. Circulation. 2012;125(15):1848–1857. , , , et al.
- Heart disease and stroke statistics—2012 update: a report from the American Heart Association. Circulation. 2012;125(1):e2–e220. , , , et al.
- ACC/AHA 2007 guidelines for the management of patients with unstable angina/non‐ST‐Elevation myocardial infarction: a report of the American College of Cardiology/American Heart Association Task Force on Practice Guidelines (Writing Committee to Revise the 2002 Guidelines for the Management of Patients With Unstable Angina/Non‐ST‐Elevation Myocardial Infarction) developed in collaboration with the American College of Emergency Physicians, the Society for Cardiovascular Angiography and Interventions, and the Society of Thoracic Surgeons endorsed by the American Association of Cardiovascular and Pulmonary Rehabilitation and the Society for Academic Emergency Medicine. J Am Coll Cardiol. 2007;50(7):e1–e157. , , , et al.
- 2007 focused update of the ACC/AHA 2004 guidelines for the management of patients with ST‐elevation myocardial infarction: a report of the American College of Cardiology/American Heart Association Task Force on Practice Guidelines. J Am Coll Cardiol. 2008;51(2):210–247. , , , et al.
- Association between adoption of evidence‐based treatment and survival for patients with ST‐elevation myocardial infarction. JAMA. 2011;305(16):1677–1684. , , , , , .
- Association of changes in clinical characteristics and management with improvement in survival among patients with ST‐elevation myocardial infarction. JAMA. 2012;308(10):998–1006. , , , et al.
- Changes in myocardial infarction guideline adherence as a function of patient risk: an end to paradoxical care? J Am Coll Cardiol. 2011;58(17):1760–1765. , , , et al.
- Care in U.S. hospitals—the Hospital Quality Alliance program. N Engl J Med. 2005;353(3):265–274. , , , .
- Challenges in the treatment of NSTEMI patients at high risk for both ischemic and bleeding events: insights from the ACTION Registry‐GWTG. J Am Coll Cardiol. 2011;57:E913–E913. , , et al.
- Quality of care in U.S. hospitals as reflected by standardized measures, 2002–2004. N Engl J Med. 2005;353(3):255–264. , , , , .
- Guideline‐based standardized care is associated with substantially lower mortality in medicare patients with acute myocardial infarction. J Am Coll Cardiol. 2005;46(7):1242–1248. , , .
- Impact of a standardized heart failure order set on mortality, readmission, and quality and costs of care. Int J Qual Health Care. 2010;22(6):437–444. , , , et al.
- Linking automated databases for research in managed care settings. Ann Intern Med. 1997;127(8 pt 2):719–724. .
- Risk adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , , , .
- Length of stay predictions: improvements through the use of automated laboratory and comorbidity variables. Med Care. 2010;48(8):739–744. , , , .
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6(2):74–80. , , , , , .
- Adverse outcomes associated with delayed intensive care unit transfers in an integrated healthcare system. J Hosp Med. 2012;7(3):224–230. , , , .
- International Classification of Diseases, 9th Revision‐Clinical Modification. 4th ed. 3 Vols. Los Angeles, CA: Practice Management Information Corporation; 2006.
- Anticoagulation therapy for stroke prevention in atrial fibrillation: how well do randomized trials translate into clinical practice? JAMA. 2003;290(20):2685–2692. , , , et al.
- Early detection of impending physiologic deterioration among patients who are not in intensive care: development of predictive models using data from an automated electronic medical record. J Hosp Med. 2012;7(5):388–395. , , , , , .
- Risk‐adjusting hospital mortality using a comprehensive electronic record in an integrated healthcare delivery system. Med Care. 2013;51(5):446–453. , , , , .
- Effect of choice of estimation method on inter‐hospital mortality rate comparisons. Med Care. 2010;48(5):456–485. , , .
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population. J Clin Epidemiol. 2010;63(7):798–803. , , , .
- Derivation and validation of a model to predict daily risk of death in hospital. Med Care. 2011;49(8):734–743. , , , , .
- Adapting a clinical comorbidity index for use with ICD‐9‐CM administrative databases. J Clin Epidemiol. 1992;45(6):613–619. , , .
- Introduction to Statistical Mediation Analysis. New York, NY: Lawrence Erlbaum Associates; 2008. .
- Nonparametric estimation of average treatment effects under exogenity: a review. Rev Econ Stat. 2004;86:25. .
- Design of Observational Studies. New York, NY: Springer Science+Business Media; 2010. .
- Balance diagnostics for comparing the distribution of baseline covariates between treatment groups in propensity‐score matched samples. Stat Med. 2009;28:24. .
- Estimation of regression coefficients when some regressors are not always observed. J Am Stat Assoc. 1994(89):846–866. , , .
- Stratification and weighting via the propensity score in estimation of causal treatment effects: a comparative study. Stat Med. 2004;23(19):2937–2960. , .
- Discussing hidden bias in observational studies. Ann Intern Med. 1991;115(11):901–905. .
- Propensity score methods for bias reduction in the comparison of a treatment to a non‐randomized control group. Stat Med. 1998;17(19):2265–2281. .
- A method/macro based on propensity score and Mahalanobis distance to reduce bias in treatment comparison in observational study, 2005. www.lexjansen.com/pharmasug/2006/publichealthresearch/pr05.pdf. Accessed on September 14, 2013. , , .
- Using health information technology to improve health care. Arch Intern Med. 2012;172(22):1728–1730. .
© 2014 Society of Hospital Medicine
Risk Factors For Unplanned ICU Transfer
Emergency Department (ED) patients who are hospitalized and require unplanned transfer to the intensive care unit (ICU) within 24 hours of arrival on the ward have higher mortality than direct ICU admissions.1, 2 Previous research found that 5% of ED admissions experienced unplanned ICU transfer during their hospitalization, yet these patients account for 25% of in‐hospital deaths and have a longer length of stay than direct ICU admissions.1, 3 For these reasons, inpatient rapid‐response teams and early warning systems have been studied to reduce the mortality of patients who rapidly deteriorate on the hospital ward.410 However, there is little conclusive evidence that these interventions decrease mortality.710 It is possible that with better recognition and intervention in the ED, a portion of these unplanned ICU transfers and their subsequent adverse outcomes could be prevented.11
Previous research on risk factors for unplanned ICU transfers among ED admissions is limited. While 2 previous studies from non‐US hospitals used administrative data to identify some general populations at risk for unplanned ICU transfer,12, 13 these studies did not differentiate between transfers shortly after admission and those that occurred during a prolonged hospital staya critical distinction since the outcomes between these groups differs substantially.1 Another limitation of these studies is the absence of physiologic measures at ED presentation, which have been shown to be highly predictive of mortality.14
In this study, we describe risk factors for unplanned transfer to the ICU within 24 hours of arrival on the ward, among a large cohort of ED hospitalizations across 13 community hospitals. Focusing on admitting diagnoses most at risk, our goal was to inform efforts to improve the triage of ED admissions and determine which patients may benefit from additional interventions, such as improved resuscitation, closer monitoring, or risk stratification tools. We also hypothesized that higher volume hospitals would have lower rates of unplanned ICU transfers, as these hospitals are more likely have more patient care resources on the hospital ward and a higher threshold to transfer to the ICU.
METHODS
Setting and Patients
The setting for this study was Kaiser Permanente Northern California (KPNC), a large integrated healthcare delivery system serving approximately 3.3 million members.1, 3, 15, 16 We extracted data on all adult ED admissions (18 years old) to the hospital between 2007 and 2009. We excluded patients who went directly to the operating room or the ICU, as well as gynecological/pregnancy‐related admissions, as these patients have substantially different mortality risks.14 ED admissions to hospital wards could either go to medicalsurgical units or transitional care units (TCU), an intermediate level of care between the medicalsurgical units and the ICU. We chose to focus on hospitals with similar inpatient structures. Thus, 8 hospitals without TCUs were excluded, leaving 13 hospitals for analysis. The KPNC Institutional Review Board approved this study.
Main Outcome Measure
The main outcome measure was unplanned transfer to the ICU within 24 hours of arrival to the hospital ward, based upon bed history data. As in previous research, we make the assumptionwhich is supported by the high observed‐to‐expected mortality ratios found in these patientsthat these transfers to the ICU were due to clinical deterioration, and thus were unplanned, rather than a planned transfer to the ICU as is more common after an elective surgical procedure.13 The comparison population was patients admitted from the ED to the ward who never experienced a transfer to the ICU.
Patient and Hospital Characteristics
We extracted patient data on age, sex, admitting diagnosis, chronic illness burden, acute physiologic derangement in the ED, and hospital unit length of stay. Chronic illness was measured using the Comorbidity Point Score (COPS), and physiologic derangement was measured using the Laboratory Acute Physiology Score (LAPS) calculated from labs collected in the ED.1, 14, 17 The derivation of these variables from the electronic medical record has been previously described.14 The COPS was derived from International Classification of Diseases, Ninth Revision (ICD‐9) codes for all Kaiser Permanente Medical Care Program (KPMCP) inpatient and outpatient encounters prior to hospitalization. The LAPS is based on 14 possible lab tests that could be drawn in the ED or in the 72 hours prior to hospitalization. The admitting diagnosis is the ICD‐9 code assigned for the primary diagnosis determined by the admitting physician at the time when hospital admission orders are entered. We further collapsed a previously used categorization of 44 primary condition diagnoses, based on admission ICD‐9 codes,14 into 25 broad diagnostic categories based on pathophysiologic plausibility and mortality rates. We tabulated inpatient admissions originating in the ED to derive a hospital volume measure.
Statistical Analyses
We compared patient characteristics, hospital volume, and outcomes by whether or not an unplanned ICU transfer occurred. Unadjusted analyses were performed with analysis of variance (ANOVA) and chi‐square tests. We calculated crude rates of unplanned ICU transfer per 1,000 ED inpatient admissions by patient characteristics and by hospital, stratified by hospital volume.
We used a hierarchical multivariate logistic regression model to estimate adjusted odds ratios for unplanned ICU transfer as a function of both patient‐level variables (age, sex, COPS, LAPS, time of admission, admission to TCU vs ward, admitting diagnosis) and hospital‐level variables (volume) in the model. We planned to choose the reference group for admitting diagnosis as the one with an unadjusted odds ratio closest to the null (1.00). This model addresses correlations between patients with multiple hospitalizations and clustering by hospital, by fitting random intercepts for these clusters. All analyses were performed in Stata 12 (StataCorp, College Station, TX), and statistics are presented with 95% confidence intervals (CI). The Stata program gllamm (Generalized Linear Latent and Mixed Models) was used for hierarchical modeling.18
RESULTS
Of 178,315 ED non‐ICU hospitalizations meeting inclusion criteria, 4,252 (2.4%) were admitted to the ward and were transferred to the ICU within 24 hours of leaving the ED. There were 122,251 unique patients in our study population. Table 1 compares the characteristics of ED hospitalizations in which an unplanned transfer occurred to those that did not experience an unplanned transfer. Unplanned transfers were more likely to have a higher comorbidity burden, more deranged physiology, and more likely to arrive on the floor during the overnight shift.
Characteristics | Unplanned Transfer to ICU Within 24 h of Leaving ED? | P Value* | |
---|---|---|---|
Yes | No | ||
N = 4,252 (2.4%) | N = 174,063 (97.6%) | ||
| |||
Age, median (IQR) | 69 (5680) | 70 (5681) | <0.01 |
Male, % | 51.3 | 45.9 | <0.01 |
Comorbidity Points Score (COPS), median (IQR) | 100 (46158) | 89 (42144) | <0.01 |
Laboratory Acute Physiology Score (LAPS), median (IQR) | 26 (1342) | 18 (633) | <0.01 |
Nursing shift on arrival to floor, % | |||
Day: 7 am3 pm (Reference) | 20.1 | 20.1 | NS |
Evening: 3 pm11 pm | 47.6 | 50.2 | NS |
Overnight: 11 pm7 am | 32.3 | 29.7 | <0.01 |
Weekend admission, % | 33.7 | 32.7 | NS |
Admitted to monitored bed, % | 24.1 | 24.9 | NS |
Emergency department annual volume, mean (SD) | 48,755 (15,379) | 50,570 (15,276) | <0.01 |
Non‐ICU annual admission volume, mean (SD) | 5,562 (1,626) | 5,774 (1,568) | <0.01 |
Admitting diagnosis, listed by descending frequency, % | NS | ||
Pneumonia and respiratory infections | 16.3 | 11.8 | <0.01 |
Gastrointestinal bleeding | 12.8 | 13.6 | NS |
Chest pain | 7.3 | 10.0 | <0.01 |
Miscellaneous conditions | 5.6 | 6.2 | NS |
All other acute infections | 4.7 | 6.0 | <0.01 |
Seizures | 4.1 | 5.9 | <0.01 |
AMI | 3.9 | 3.3 | <0.05 |
COPD | 3.8 | 3.0 | <0.01 |
CHF | 3.5 | 3.7 | NS |
Arrhythmias and pulmonary embolism | 3.5 | 3.3 | NS |
Stroke | 3.4 | 3.5 | NS |
Diabetic emergencies | 3.3 | 2.6 | <0.01 |
Metabolic, endocrine, electrolytes | 3.0 | 2.9 | NS |
Sepsis | 3.0 | 1.2 | <0.01 |
Other neurology and toxicology | 3.0 | 2.9 | NS |
Urinary tract infections | 2.9 | 3.2 | NS |
Catastrophic conditions | 2.6 | 1.2 | <0.01 |
Rheumatology | 2.5 | 3.5 | <0.01 |
Hematology and oncology | 2.4 | 2.4 | NS |
Acute renal failure | 1.9 | 1.1 | <0.01 |
Pancreatic and liver | 1.7 | 2.0 | NS |
Trauma, fractures, and dislocations | 1.6 | 1.8 | NS |
Bowel obstructions and diseases | 1.6 | 2.9 | <0.01 |
Other cardiac conditions | 1.5 | 1.3 | NS |
Other renal conditions | 0.6 | 1.0 | <0.01 |
Inpatient length of stay, median days (IQR) | 4.7 (2.78.6) | 2.6 (1.54.4) | <0.01 |
Died during hospitalization, % | 12.7 | 2.4 | <0.01 |
Unplanned ICU transfers were more frequent in lower volume hospitals (Table 1). Figure 1 displays the inverse relationship between hospital annual ED inpatient admission volume and unplanned ICU transfers rates. The lowest volume hospital had a crude rate twice as high as the 2 highest volume hospitals (39 vs 20, per 1,000 admissions).
Pneumonia/respiratory infection was the most frequent admitting condition associated with unplanned transfer (16.3%) (Table 1). There was also wide variation in crude rates for unplanned ICU transfer by admitting condition (Figure 2). Patients admitted with sepsis had the highest rate (59 per 1,000 admissions), while patients admitted with renal conditions other than acute renal failure had the lowest rates (14.3 per 1,000 admissions).
We confirmed that almost all diagnoses found to account for a disproportionately high share of unplanned ICU transfers in Table 1 were indeed independently associated with this phenomenon after adjustment for patient and hospital differences (Figure 2). Pneumonia remained the most frequent condition associated with unplanned ICU transfer (odds ratio [OR] 1.50; 95% CI 1.201.86). Although less frequent, sepsis had the strongest association of any condition with unplanned transfer (OR 2.51; 95% CI 1.903.31). However, metabolic, endocrine, and electrolyte conditions were no longer associated with unplanned transfer after adjustment, while arrhythmias and pulmonary embolism were. Other conditions confirmed to be associated with increased risk of unplanned transfer included: myocardial infarction (MI), chronic obstructive pulmonary disease (COPD), stroke, diabetic emergencies, catastrophic conditions (includes aortic catastrophes, all forms of shock except septic shock, and intracranial hemorrhage), and acute renal failure. After taking into account the frequency of admitting diagnoses, respiratory conditions (COPD, pneumonia/acute respiratory infection) comprised nearly half (47%) of all conditions associated with increased risk of unplanned ICU transfer.
Other factors confirmed to be independently associated with unplanned ICU transfer included: male sex (OR 1.20; 95% CI 1.131.28), high comorbidity burden as measured by COPS >145 (OR 1.13; 95% CI 1.031.24), increasingly abnormal physiology compared to a LAPS <7, and arrival on ward during the overnight shift (OR 1.10; 95% CI 1.011.21). After adjustment, we did find that admission to the TCU rather than a medicalsurgical unit was associated with decreased risk of unplanned ICU transfer (OR 0.83; 95% CI 0.770.90). Age 85 was associated with decreased risk of unplanned ICU transfer relative to the youngest age group of 1834‐year‐old patients (OR 0.64; 95% CI 0.530.77).
ED admissions to higher volume hospitals were 6% less likely to experience an unplanned transfer for each additional 1,000 annual ED hospitalizations over a lower volume hospital (OR 0.94; 95% CI 0.910.98). In other words, a patient admitted to a hospital with 8,000 annual ED hospitalizations had 30% decreased odds of unplanned ICU transfer compared to a hospital with only 3,000 annual ED hospitalizations.
DISCUSSION
Patients admitted with respiratory conditions accounted for half of all admitting diagnoses associated with increased risk of unplanned transfer to the ICU within 24 hours of arrival to the ward. We found that 1 in 30 ED ward admissions for pneumonia, and 1 in 33 for COPD, were transferred to the ICU within 24 hours. These findings indicate that there is some room for improvement in early care of respiratory conditions, given the average unplanned transfer rate of 1 in 42, and previous research showing that patients with pneumonia and patients with COPD, who experience unplanned ICU transfer, have substantially worse mortality than those directly admitted to the ICU.1
Although less frequent than hospitalizations for respiratory conditions, patients admitted with sepsis were at the highest risk of unplanned ICU transfer (1 in 17 ED non‐ICU hospitalizations). We also found that MI and stroke ward admissions had a higher risk of unplanned ICU transfer. However, we previously found that unplanned ICU transfers for sepsis, MI, and stroke did not have worse mortality than direct ICU admits for these conditions.1 Therefore, quality improvement efforts to reduce excess mortality related to early decompensation in the hospital and unplanned ICU transfer would be most effective if targeted towards respiratory conditions such as pneumonia and COPD.
This is the only in‐depth study, to our knowledge, to explore the association between a set of mutually exclusive diagnostic categories and risk of unplanned ICU transfer within 24 hours, and it is the first study to identify risk factors for unplanned ICU transfer in a multi‐hospital cohort adjusted for patient‐ and hospital‐level characteristics. We also identified a novel hospital volumeoutcome relationship: Unplanned ICU transfers are up to twice as likely to occur in the smallest volume hospitals compared with highest volume hospitals. Hospital volume has long been proposed as a proxy for hospital resources; there are several studies showing a relationship between low‐volume hospitals and worse outcomes for a number of conditions.19, 20 Possible mechanisms may include decreased ICU capacity, decreased on‐call intensivists in the hospital after hours, and less experience with certain critical care conditions seen more frequently in high‐volume hospitals.21
Patients at risk of unplanned ICU transfer were also more likely to have physiologic derangement identified on laboratory testing, high comorbidity burden, and arrive on the ward between 11 PM and 7 AM. Given the strong correlation between comorbidity burden and physiologic derangement and mortality,14 it is not surprising that the COPS and LAPS were independent predictors of unplanned transfer. It is unclear, however, why arriving on the ward on the overnight shift is associated with higher risk. One possibility is that patients who arrive on the wards during 11 PM to 7 AM are also likely to have been in the ED during evening peak hours most associated with ED crowding.22 High levels of ED crowding have been associated with delays in care, worse quality care, lapses in patient safety, and even increased in‐hospital mortality.22, 23 Other possible reasons include decreased in‐hospital staffing and longer delays in critical diagnostic tests and interventions.2428
Admission to TCUs was associated with decreased risk of unplanned ICU transfer in the first 24 hours of hospitalization. This may be due to the continuous monitoring, decreased nursing‐to‐patient ratios, or the availability to provide some critical care interventions. In our study, age 85 was associated with lower likelihood of unplanned transfer. Unfortunately, we did not have access to data on advanced directives or patient preferences. Data on advanced directives would help to distinguish whether this phenomenon was related to end‐of‐life care goals versus other explanations.
Our study confirms some risk factors identified in previous studies. These include specific diagnoses such as pneumonia and COPD,12, 13, 29 heavy comorbidity burden,12, 13, 29 abnormal labs,29 and male sex.13 Pneumonia has consistently been shown to be a risk factor for unplanned ICU transfer. This may stem from the dynamic nature of this condition and its ability to rapidly progress, and the fact that some ICUs may not accept pneumonia patients unless they demonstrate a need for mechanical ventilation.30 Recently, a prediction rule has been developed to determine which patients with pneumonia are likely to have an unplanned ICU transfer.30 It is possible that with validation and application of this rule, unplanned transfer rates for pneumonia could be reduced. It is unclear whether males have unmeasured factors associated with increased risk of unplanned transfer or whether a true gender disparity exists.
Our findings should be interpreted within the context of this study's limitations. First, this study was not designed to distinguish the underlying cause of the unplanned transfer such as under‐recognition of illness severity in the ED, evolving clinical disease after leaving the ED, or delays in critical interventions on the ward. These are a focus of our ongoing research efforts. Second, while previous studies have demonstrated that our automated risk adjustment variables can accurately predict in‐hospital mortality (0.88 area under curve in external populations),17 additional data on vital signs and mental status could further improve risk adjustment. However, using automated data allowed us to study risk factors for unplanned transfer in a multi‐hospital cohort with a much larger population than has been previously studied. Serial data on vital signs and mental status both in the ED and during hospitalization could also be helpful in determining which unplanned transfers could be prevented with earlier recognition and intervention. Finally, all patient care occurred within an integrated healthcare delivery system. Thus, differences in case‐mix, hospital resources, ICU structure, and geographic location should be considered when applying our results to other healthcare systems.
This study raises several new areas for future research. With access to richer data becoming available in electronic medical records, prediction rules should be developed to enable better triage to appropriate levels of care for ED admissions. Future research should also analyze the comparative effectiveness of intermediate monitored units versus non‐monitored wards for preventing clinical deterioration by admitting diagnosis. Diagnoses that have been shown to have an increased risk of death after unplanned ICU transfer, such as pneumonia/respiratory infection and COPD,1 should be prioritized in this research. Better understanding is needed on the diagnosis‐specific differences and the differences in ED triage process and ICU structure that may explain why high‐volume hospitals have significantly lower rates of early unplanned ICU transfers compared with low‐volume hospitals. In particular, determining the effect of TCU and ICU capacities and census at the time of admission, and comparing patient risk characteristics across hospital‐volume strata would be very useful. Finally, more work is needed to determine whether the higher rate of unplanned transfers during overnight nursing shifts is related to decreased resource availability, preceding ED crowding, or other organizational causes.
In conclusion, patients admitted with respiratory conditions, sepsis, MI, high comorbidity, and abnormal labs are at modestly increased risk of unplanned ICU transfer within 24 hours of admission from the ED. Patients admitted with respiratory conditions (pneumonia/respiratory infections and COPD) accounted for half of the admitting diagnoses that are at increased risk for unplanned ICU transfer. These patients may benefit from better inpatient triage from the ED, earlier intervention, or closer monitoring. More research is needed to determine the specific aspects of care associated with admission to intermediate care units and high‐volume hospitals that reduce the risk of unplanned ICU transfer.
Acknowledgements
The authors thank John D. Greene, Juan Carlos La Guardia, and Benjamin Turk for their assistance with formatting of the dataset; Dr Alan S. Go, Acting Director of the Division of Research, for reviewing the manuscript; and Alina Schnake‐Mahl for formatting the manuscript.
- Adverse outcomes associated with delayed intensive care unit transfers in an integrated healthcare system. J Hosp Med. 2011;7(3):224–230. , , , et al.
- Inpatient transfers to the intensive care unit. J Gen Intern Med. 2003;18(2):77–83. , , , et al.
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6:74–80. , , , et al.
- Hospital‐wide code rates and mortality before and after implementation of a rapid response team. JAMA. 2008;300(21):2506–2513. , , , et al.
- Effect of a rapid response team on hospital‐wide mortality and code rates outside the ICU in a children's hospital. JAMA. 2007;298(19):2267–2274. , , , et al.
- Introduction of the medical emergency team (MET) system: a cluster‐randomised controlled trial. Lancet. 2005;365(9477):2091–2097. , , , et al.
- Rapid response systems: A systematic review. Crit Care Med. 2007;35(5):1238–1243. , , , et al.
- Effects of rapid response systems on clinical outcomes: systematic review and meta‐analysis. J Hosp Med. 2007;2(6):422–432. , , , et al.
- Rapid response teams: a systematic review and meta‐analysis. Arch Intern Med. 2010;170(1):18–26. , , , et al.
- Outreach and early warning systems (EWS) for the prevention of intensive care admission and death of critically ill adult patients on general hospital wards. Cochrane Database Syst Rev. 2007;3:CD005529. , , , et al.
- Unplanned transfers to a medical intensive care unit: Causes and relationship to preventable errors in care. J Hosp Med. 2011;6:68–72. , , , et al.
- Using administrative data to develop a nomogram for individualising risk of unplanned admission to intensive care. Resuscitation. 2008;79(2):241–248. , , , et al.
- Unplanned admission to intensive care after emergency hospitalisation: risk factors and development of a nomogram for individualising risk. Resuscitation. 2009;80(2):224–230. , , , et al.
- Risk‐adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , et al.
- Linking automated databases for research in managed care settings. Ann Intern Med. 1997;127(8 pt 2):719–724. .
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases. Am J Manag Care. 2008;14(3):158–166. , , , et al.
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population. J Clin Epidemiol. 2011;63(7):798–803. , , , et al.
- Maximum likelihood estimation of limited and discrete dependent variable models with nested random effects. J Econometrics. 2005;128(2):301–323. , , .
- The relation between volume and outcome in health care. N Engl J Med. 1999;340(21):1677–1679. .
- Is volume related to outcome in health care? A systematic review and methodologic critique of the literature. Ann Intern Med. 2002;137(6):511–520. , , .
- Working with capacity limitations: operations management in critical care. Crit Care. 2011;15(4):308. , , .
- Systematic review of emergency department crowding: causes, effects, and solutions. Ann Intern Med. 2008;52(2):126–136. , .
- The effect of emergency department crowding on clinically oriented outcomes. Acad Emerg Med. 2009;16(1):1–10. , , , et al.
- Association between time of admission to the ICU and mortality. Chest. 2010;138(1):68–75. , , , et al.
- Off‐hour admission and in‐hospital stroke case fatality in the get with the guidelines‐stroke program. Stroke. 2009;40(2):569–576. , , , et al.
- Relationship between time of day, day of week, timeliness of reperfusion, and in‐hospital mortality for patients with acute ST‐segment elevation myocardial infarction. JAMA. 2005;294(7):803–812. , , , et al.
- Hospital mortality among adults admitted to and discharged from intensive care on weekends and evenings. J Crit Care. 2008;23(3):317–324. , , , et al.
- Association between ICU admission during morning rounds and mortality. Chest. 2009;136(6):1489–1495. , , , et al.
- Identifying infected emergency department patients admitted to the hospital ward at risk of clinical deterioration and intensive care unit transfer. Acad Emerg Med. 2010;17(10):1080–1085. , , , et al.
- Risk stratification of early admission to the intensive care unit of patients with no major criteria of severe community‐acquired pneumonia: development of an international prediction rule. Crit Care. 2009;13(2):R54. , , , et al.
Emergency Department (ED) patients who are hospitalized and require unplanned transfer to the intensive care unit (ICU) within 24 hours of arrival on the ward have higher mortality than direct ICU admissions.1, 2 Previous research found that 5% of ED admissions experienced unplanned ICU transfer during their hospitalization, yet these patients account for 25% of in‐hospital deaths and have a longer length of stay than direct ICU admissions.1, 3 For these reasons, inpatient rapid‐response teams and early warning systems have been studied to reduce the mortality of patients who rapidly deteriorate on the hospital ward.410 However, there is little conclusive evidence that these interventions decrease mortality.710 It is possible that with better recognition and intervention in the ED, a portion of these unplanned ICU transfers and their subsequent adverse outcomes could be prevented.11
Previous research on risk factors for unplanned ICU transfers among ED admissions is limited. While 2 previous studies from non‐US hospitals used administrative data to identify some general populations at risk for unplanned ICU transfer,12, 13 these studies did not differentiate between transfers shortly after admission and those that occurred during a prolonged hospital staya critical distinction since the outcomes between these groups differs substantially.1 Another limitation of these studies is the absence of physiologic measures at ED presentation, which have been shown to be highly predictive of mortality.14
In this study, we describe risk factors for unplanned transfer to the ICU within 24 hours of arrival on the ward, among a large cohort of ED hospitalizations across 13 community hospitals. Focusing on admitting diagnoses most at risk, our goal was to inform efforts to improve the triage of ED admissions and determine which patients may benefit from additional interventions, such as improved resuscitation, closer monitoring, or risk stratification tools. We also hypothesized that higher volume hospitals would have lower rates of unplanned ICU transfers, as these hospitals are more likely have more patient care resources on the hospital ward and a higher threshold to transfer to the ICU.
METHODS
Setting and Patients
The setting for this study was Kaiser Permanente Northern California (KPNC), a large integrated healthcare delivery system serving approximately 3.3 million members.1, 3, 15, 16 We extracted data on all adult ED admissions (18 years old) to the hospital between 2007 and 2009. We excluded patients who went directly to the operating room or the ICU, as well as gynecological/pregnancy‐related admissions, as these patients have substantially different mortality risks.14 ED admissions to hospital wards could either go to medicalsurgical units or transitional care units (TCU), an intermediate level of care between the medicalsurgical units and the ICU. We chose to focus on hospitals with similar inpatient structures. Thus, 8 hospitals without TCUs were excluded, leaving 13 hospitals for analysis. The KPNC Institutional Review Board approved this study.
Main Outcome Measure
The main outcome measure was unplanned transfer to the ICU within 24 hours of arrival to the hospital ward, based upon bed history data. As in previous research, we make the assumptionwhich is supported by the high observed‐to‐expected mortality ratios found in these patientsthat these transfers to the ICU were due to clinical deterioration, and thus were unplanned, rather than a planned transfer to the ICU as is more common after an elective surgical procedure.13 The comparison population was patients admitted from the ED to the ward who never experienced a transfer to the ICU.
Patient and Hospital Characteristics
We extracted patient data on age, sex, admitting diagnosis, chronic illness burden, acute physiologic derangement in the ED, and hospital unit length of stay. Chronic illness was measured using the Comorbidity Point Score (COPS), and physiologic derangement was measured using the Laboratory Acute Physiology Score (LAPS) calculated from labs collected in the ED.1, 14, 17 The derivation of these variables from the electronic medical record has been previously described.14 The COPS was derived from International Classification of Diseases, Ninth Revision (ICD‐9) codes for all Kaiser Permanente Medical Care Program (KPMCP) inpatient and outpatient encounters prior to hospitalization. The LAPS is based on 14 possible lab tests that could be drawn in the ED or in the 72 hours prior to hospitalization. The admitting diagnosis is the ICD‐9 code assigned for the primary diagnosis determined by the admitting physician at the time when hospital admission orders are entered. We further collapsed a previously used categorization of 44 primary condition diagnoses, based on admission ICD‐9 codes,14 into 25 broad diagnostic categories based on pathophysiologic plausibility and mortality rates. We tabulated inpatient admissions originating in the ED to derive a hospital volume measure.
Statistical Analyses
We compared patient characteristics, hospital volume, and outcomes by whether or not an unplanned ICU transfer occurred. Unadjusted analyses were performed with analysis of variance (ANOVA) and chi‐square tests. We calculated crude rates of unplanned ICU transfer per 1,000 ED inpatient admissions by patient characteristics and by hospital, stratified by hospital volume.
We used a hierarchical multivariate logistic regression model to estimate adjusted odds ratios for unplanned ICU transfer as a function of both patient‐level variables (age, sex, COPS, LAPS, time of admission, admission to TCU vs ward, admitting diagnosis) and hospital‐level variables (volume) in the model. We planned to choose the reference group for admitting diagnosis as the one with an unadjusted odds ratio closest to the null (1.00). This model addresses correlations between patients with multiple hospitalizations and clustering by hospital, by fitting random intercepts for these clusters. All analyses were performed in Stata 12 (StataCorp, College Station, TX), and statistics are presented with 95% confidence intervals (CI). The Stata program gllamm (Generalized Linear Latent and Mixed Models) was used for hierarchical modeling.18
RESULTS
Of 178,315 ED non‐ICU hospitalizations meeting inclusion criteria, 4,252 (2.4%) were admitted to the ward and were transferred to the ICU within 24 hours of leaving the ED. There were 122,251 unique patients in our study population. Table 1 compares the characteristics of ED hospitalizations in which an unplanned transfer occurred to those that did not experience an unplanned transfer. Unplanned transfers were more likely to have a higher comorbidity burden, more deranged physiology, and more likely to arrive on the floor during the overnight shift.
Characteristics | Unplanned Transfer to ICU Within 24 h of Leaving ED? | P Value* | |
---|---|---|---|
Yes | No | ||
N = 4,252 (2.4%) | N = 174,063 (97.6%) | ||
| |||
Age, median (IQR) | 69 (5680) | 70 (5681) | <0.01 |
Male, % | 51.3 | 45.9 | <0.01 |
Comorbidity Points Score (COPS), median (IQR) | 100 (46158) | 89 (42144) | <0.01 |
Laboratory Acute Physiology Score (LAPS), median (IQR) | 26 (1342) | 18 (633) | <0.01 |
Nursing shift on arrival to floor, % | |||
Day: 7 am3 pm (Reference) | 20.1 | 20.1 | NS |
Evening: 3 pm11 pm | 47.6 | 50.2 | NS |
Overnight: 11 pm7 am | 32.3 | 29.7 | <0.01 |
Weekend admission, % | 33.7 | 32.7 | NS |
Admitted to monitored bed, % | 24.1 | 24.9 | NS |
Emergency department annual volume, mean (SD) | 48,755 (15,379) | 50,570 (15,276) | <0.01 |
Non‐ICU annual admission volume, mean (SD) | 5,562 (1,626) | 5,774 (1,568) | <0.01 |
Admitting diagnosis, listed by descending frequency, % | NS | ||
Pneumonia and respiratory infections | 16.3 | 11.8 | <0.01 |
Gastrointestinal bleeding | 12.8 | 13.6 | NS |
Chest pain | 7.3 | 10.0 | <0.01 |
Miscellaneous conditions | 5.6 | 6.2 | NS |
All other acute infections | 4.7 | 6.0 | <0.01 |
Seizures | 4.1 | 5.9 | <0.01 |
AMI | 3.9 | 3.3 | <0.05 |
COPD | 3.8 | 3.0 | <0.01 |
CHF | 3.5 | 3.7 | NS |
Arrhythmias and pulmonary embolism | 3.5 | 3.3 | NS |
Stroke | 3.4 | 3.5 | NS |
Diabetic emergencies | 3.3 | 2.6 | <0.01 |
Metabolic, endocrine, electrolytes | 3.0 | 2.9 | NS |
Sepsis | 3.0 | 1.2 | <0.01 |
Other neurology and toxicology | 3.0 | 2.9 | NS |
Urinary tract infections | 2.9 | 3.2 | NS |
Catastrophic conditions | 2.6 | 1.2 | <0.01 |
Rheumatology | 2.5 | 3.5 | <0.01 |
Hematology and oncology | 2.4 | 2.4 | NS |
Acute renal failure | 1.9 | 1.1 | <0.01 |
Pancreatic and liver | 1.7 | 2.0 | NS |
Trauma, fractures, and dislocations | 1.6 | 1.8 | NS |
Bowel obstructions and diseases | 1.6 | 2.9 | <0.01 |
Other cardiac conditions | 1.5 | 1.3 | NS |
Other renal conditions | 0.6 | 1.0 | <0.01 |
Inpatient length of stay, median days (IQR) | 4.7 (2.78.6) | 2.6 (1.54.4) | <0.01 |
Died during hospitalization, % | 12.7 | 2.4 | <0.01 |
Unplanned ICU transfers were more frequent in lower volume hospitals (Table 1). Figure 1 displays the inverse relationship between hospital annual ED inpatient admission volume and unplanned ICU transfers rates. The lowest volume hospital had a crude rate twice as high as the 2 highest volume hospitals (39 vs 20, per 1,000 admissions).
Pneumonia/respiratory infection was the most frequent admitting condition associated with unplanned transfer (16.3%) (Table 1). There was also wide variation in crude rates for unplanned ICU transfer by admitting condition (Figure 2). Patients admitted with sepsis had the highest rate (59 per 1,000 admissions), while patients admitted with renal conditions other than acute renal failure had the lowest rates (14.3 per 1,000 admissions).
We confirmed that almost all diagnoses found to account for a disproportionately high share of unplanned ICU transfers in Table 1 were indeed independently associated with this phenomenon after adjustment for patient and hospital differences (Figure 2). Pneumonia remained the most frequent condition associated with unplanned ICU transfer (odds ratio [OR] 1.50; 95% CI 1.201.86). Although less frequent, sepsis had the strongest association of any condition with unplanned transfer (OR 2.51; 95% CI 1.903.31). However, metabolic, endocrine, and electrolyte conditions were no longer associated with unplanned transfer after adjustment, while arrhythmias and pulmonary embolism were. Other conditions confirmed to be associated with increased risk of unplanned transfer included: myocardial infarction (MI), chronic obstructive pulmonary disease (COPD), stroke, diabetic emergencies, catastrophic conditions (includes aortic catastrophes, all forms of shock except septic shock, and intracranial hemorrhage), and acute renal failure. After taking into account the frequency of admitting diagnoses, respiratory conditions (COPD, pneumonia/acute respiratory infection) comprised nearly half (47%) of all conditions associated with increased risk of unplanned ICU transfer.
Other factors confirmed to be independently associated with unplanned ICU transfer included: male sex (OR 1.20; 95% CI 1.131.28), high comorbidity burden as measured by COPS >145 (OR 1.13; 95% CI 1.031.24), increasingly abnormal physiology compared to a LAPS <7, and arrival on ward during the overnight shift (OR 1.10; 95% CI 1.011.21). After adjustment, we did find that admission to the TCU rather than a medicalsurgical unit was associated with decreased risk of unplanned ICU transfer (OR 0.83; 95% CI 0.770.90). Age 85 was associated with decreased risk of unplanned ICU transfer relative to the youngest age group of 1834‐year‐old patients (OR 0.64; 95% CI 0.530.77).
ED admissions to higher volume hospitals were 6% less likely to experience an unplanned transfer for each additional 1,000 annual ED hospitalizations over a lower volume hospital (OR 0.94; 95% CI 0.910.98). In other words, a patient admitted to a hospital with 8,000 annual ED hospitalizations had 30% decreased odds of unplanned ICU transfer compared to a hospital with only 3,000 annual ED hospitalizations.
DISCUSSION
Patients admitted with respiratory conditions accounted for half of all admitting diagnoses associated with increased risk of unplanned transfer to the ICU within 24 hours of arrival to the ward. We found that 1 in 30 ED ward admissions for pneumonia, and 1 in 33 for COPD, were transferred to the ICU within 24 hours. These findings indicate that there is some room for improvement in early care of respiratory conditions, given the average unplanned transfer rate of 1 in 42, and previous research showing that patients with pneumonia and patients with COPD, who experience unplanned ICU transfer, have substantially worse mortality than those directly admitted to the ICU.1
Although less frequent than hospitalizations for respiratory conditions, patients admitted with sepsis were at the highest risk of unplanned ICU transfer (1 in 17 ED non‐ICU hospitalizations). We also found that MI and stroke ward admissions had a higher risk of unplanned ICU transfer. However, we previously found that unplanned ICU transfers for sepsis, MI, and stroke did not have worse mortality than direct ICU admits for these conditions.1 Therefore, quality improvement efforts to reduce excess mortality related to early decompensation in the hospital and unplanned ICU transfer would be most effective if targeted towards respiratory conditions such as pneumonia and COPD.
This is the only in‐depth study, to our knowledge, to explore the association between a set of mutually exclusive diagnostic categories and risk of unplanned ICU transfer within 24 hours, and it is the first study to identify risk factors for unplanned ICU transfer in a multi‐hospital cohort adjusted for patient‐ and hospital‐level characteristics. We also identified a novel hospital volumeoutcome relationship: Unplanned ICU transfers are up to twice as likely to occur in the smallest volume hospitals compared with highest volume hospitals. Hospital volume has long been proposed as a proxy for hospital resources; there are several studies showing a relationship between low‐volume hospitals and worse outcomes for a number of conditions.19, 20 Possible mechanisms may include decreased ICU capacity, decreased on‐call intensivists in the hospital after hours, and less experience with certain critical care conditions seen more frequently in high‐volume hospitals.21
Patients at risk of unplanned ICU transfer were also more likely to have physiologic derangement identified on laboratory testing, high comorbidity burden, and arrive on the ward between 11 PM and 7 AM. Given the strong correlation between comorbidity burden and physiologic derangement and mortality,14 it is not surprising that the COPS and LAPS were independent predictors of unplanned transfer. It is unclear, however, why arriving on the ward on the overnight shift is associated with higher risk. One possibility is that patients who arrive on the wards during 11 PM to 7 AM are also likely to have been in the ED during evening peak hours most associated with ED crowding.22 High levels of ED crowding have been associated with delays in care, worse quality care, lapses in patient safety, and even increased in‐hospital mortality.22, 23 Other possible reasons include decreased in‐hospital staffing and longer delays in critical diagnostic tests and interventions.2428
Admission to TCUs was associated with decreased risk of unplanned ICU transfer in the first 24 hours of hospitalization. This may be due to the continuous monitoring, decreased nursing‐to‐patient ratios, or the availability to provide some critical care interventions. In our study, age 85 was associated with lower likelihood of unplanned transfer. Unfortunately, we did not have access to data on advanced directives or patient preferences. Data on advanced directives would help to distinguish whether this phenomenon was related to end‐of‐life care goals versus other explanations.
Our study confirms some risk factors identified in previous studies. These include specific diagnoses such as pneumonia and COPD,12, 13, 29 heavy comorbidity burden,12, 13, 29 abnormal labs,29 and male sex.13 Pneumonia has consistently been shown to be a risk factor for unplanned ICU transfer. This may stem from the dynamic nature of this condition and its ability to rapidly progress, and the fact that some ICUs may not accept pneumonia patients unless they demonstrate a need for mechanical ventilation.30 Recently, a prediction rule has been developed to determine which patients with pneumonia are likely to have an unplanned ICU transfer.30 It is possible that with validation and application of this rule, unplanned transfer rates for pneumonia could be reduced. It is unclear whether males have unmeasured factors associated with increased risk of unplanned transfer or whether a true gender disparity exists.
Our findings should be interpreted within the context of this study's limitations. First, this study was not designed to distinguish the underlying cause of the unplanned transfer such as under‐recognition of illness severity in the ED, evolving clinical disease after leaving the ED, or delays in critical interventions on the ward. These are a focus of our ongoing research efforts. Second, while previous studies have demonstrated that our automated risk adjustment variables can accurately predict in‐hospital mortality (0.88 area under curve in external populations),17 additional data on vital signs and mental status could further improve risk adjustment. However, using automated data allowed us to study risk factors for unplanned transfer in a multi‐hospital cohort with a much larger population than has been previously studied. Serial data on vital signs and mental status both in the ED and during hospitalization could also be helpful in determining which unplanned transfers could be prevented with earlier recognition and intervention. Finally, all patient care occurred within an integrated healthcare delivery system. Thus, differences in case‐mix, hospital resources, ICU structure, and geographic location should be considered when applying our results to other healthcare systems.
This study raises several new areas for future research. With access to richer data becoming available in electronic medical records, prediction rules should be developed to enable better triage to appropriate levels of care for ED admissions. Future research should also analyze the comparative effectiveness of intermediate monitored units versus non‐monitored wards for preventing clinical deterioration by admitting diagnosis. Diagnoses that have been shown to have an increased risk of death after unplanned ICU transfer, such as pneumonia/respiratory infection and COPD,1 should be prioritized in this research. Better understanding is needed on the diagnosis‐specific differences and the differences in ED triage process and ICU structure that may explain why high‐volume hospitals have significantly lower rates of early unplanned ICU transfers compared with low‐volume hospitals. In particular, determining the effect of TCU and ICU capacities and census at the time of admission, and comparing patient risk characteristics across hospital‐volume strata would be very useful. Finally, more work is needed to determine whether the higher rate of unplanned transfers during overnight nursing shifts is related to decreased resource availability, preceding ED crowding, or other organizational causes.
In conclusion, patients admitted with respiratory conditions, sepsis, MI, high comorbidity, and abnormal labs are at modestly increased risk of unplanned ICU transfer within 24 hours of admission from the ED. Patients admitted with respiratory conditions (pneumonia/respiratory infections and COPD) accounted for half of the admitting diagnoses that are at increased risk for unplanned ICU transfer. These patients may benefit from better inpatient triage from the ED, earlier intervention, or closer monitoring. More research is needed to determine the specific aspects of care associated with admission to intermediate care units and high‐volume hospitals that reduce the risk of unplanned ICU transfer.
Acknowledgements
The authors thank John D. Greene, Juan Carlos La Guardia, and Benjamin Turk for their assistance with formatting of the dataset; Dr Alan S. Go, Acting Director of the Division of Research, for reviewing the manuscript; and Alina Schnake‐Mahl for formatting the manuscript.
Emergency Department (ED) patients who are hospitalized and require unplanned transfer to the intensive care unit (ICU) within 24 hours of arrival on the ward have higher mortality than direct ICU admissions.1, 2 Previous research found that 5% of ED admissions experienced unplanned ICU transfer during their hospitalization, yet these patients account for 25% of in‐hospital deaths and have a longer length of stay than direct ICU admissions.1, 3 For these reasons, inpatient rapid‐response teams and early warning systems have been studied to reduce the mortality of patients who rapidly deteriorate on the hospital ward.410 However, there is little conclusive evidence that these interventions decrease mortality.710 It is possible that with better recognition and intervention in the ED, a portion of these unplanned ICU transfers and their subsequent adverse outcomes could be prevented.11
Previous research on risk factors for unplanned ICU transfers among ED admissions is limited. While 2 previous studies from non‐US hospitals used administrative data to identify some general populations at risk for unplanned ICU transfer,12, 13 these studies did not differentiate between transfers shortly after admission and those that occurred during a prolonged hospital staya critical distinction since the outcomes between these groups differs substantially.1 Another limitation of these studies is the absence of physiologic measures at ED presentation, which have been shown to be highly predictive of mortality.14
In this study, we describe risk factors for unplanned transfer to the ICU within 24 hours of arrival on the ward, among a large cohort of ED hospitalizations across 13 community hospitals. Focusing on admitting diagnoses most at risk, our goal was to inform efforts to improve the triage of ED admissions and determine which patients may benefit from additional interventions, such as improved resuscitation, closer monitoring, or risk stratification tools. We also hypothesized that higher volume hospitals would have lower rates of unplanned ICU transfers, as these hospitals are more likely have more patient care resources on the hospital ward and a higher threshold to transfer to the ICU.
METHODS
Setting and Patients
The setting for this study was Kaiser Permanente Northern California (KPNC), a large integrated healthcare delivery system serving approximately 3.3 million members.1, 3, 15, 16 We extracted data on all adult ED admissions (18 years old) to the hospital between 2007 and 2009. We excluded patients who went directly to the operating room or the ICU, as well as gynecological/pregnancy‐related admissions, as these patients have substantially different mortality risks.14 ED admissions to hospital wards could either go to medicalsurgical units or transitional care units (TCU), an intermediate level of care between the medicalsurgical units and the ICU. We chose to focus on hospitals with similar inpatient structures. Thus, 8 hospitals without TCUs were excluded, leaving 13 hospitals for analysis. The KPNC Institutional Review Board approved this study.
Main Outcome Measure
The main outcome measure was unplanned transfer to the ICU within 24 hours of arrival to the hospital ward, based upon bed history data. As in previous research, we make the assumptionwhich is supported by the high observed‐to‐expected mortality ratios found in these patientsthat these transfers to the ICU were due to clinical deterioration, and thus were unplanned, rather than a planned transfer to the ICU as is more common after an elective surgical procedure.13 The comparison population was patients admitted from the ED to the ward who never experienced a transfer to the ICU.
Patient and Hospital Characteristics
We extracted patient data on age, sex, admitting diagnosis, chronic illness burden, acute physiologic derangement in the ED, and hospital unit length of stay. Chronic illness was measured using the Comorbidity Point Score (COPS), and physiologic derangement was measured using the Laboratory Acute Physiology Score (LAPS) calculated from labs collected in the ED.1, 14, 17 The derivation of these variables from the electronic medical record has been previously described.14 The COPS was derived from International Classification of Diseases, Ninth Revision (ICD‐9) codes for all Kaiser Permanente Medical Care Program (KPMCP) inpatient and outpatient encounters prior to hospitalization. The LAPS is based on 14 possible lab tests that could be drawn in the ED or in the 72 hours prior to hospitalization. The admitting diagnosis is the ICD‐9 code assigned for the primary diagnosis determined by the admitting physician at the time when hospital admission orders are entered. We further collapsed a previously used categorization of 44 primary condition diagnoses, based on admission ICD‐9 codes,14 into 25 broad diagnostic categories based on pathophysiologic plausibility and mortality rates. We tabulated inpatient admissions originating in the ED to derive a hospital volume measure.
Statistical Analyses
We compared patient characteristics, hospital volume, and outcomes by whether or not an unplanned ICU transfer occurred. Unadjusted analyses were performed with analysis of variance (ANOVA) and chi‐square tests. We calculated crude rates of unplanned ICU transfer per 1,000 ED inpatient admissions by patient characteristics and by hospital, stratified by hospital volume.
We used a hierarchical multivariate logistic regression model to estimate adjusted odds ratios for unplanned ICU transfer as a function of both patient‐level variables (age, sex, COPS, LAPS, time of admission, admission to TCU vs ward, admitting diagnosis) and hospital‐level variables (volume) in the model. We planned to choose the reference group for admitting diagnosis as the one with an unadjusted odds ratio closest to the null (1.00). This model addresses correlations between patients with multiple hospitalizations and clustering by hospital, by fitting random intercepts for these clusters. All analyses were performed in Stata 12 (StataCorp, College Station, TX), and statistics are presented with 95% confidence intervals (CI). The Stata program gllamm (Generalized Linear Latent and Mixed Models) was used for hierarchical modeling.18
RESULTS
Of 178,315 ED non‐ICU hospitalizations meeting inclusion criteria, 4,252 (2.4%) were admitted to the ward and were transferred to the ICU within 24 hours of leaving the ED. There were 122,251 unique patients in our study population. Table 1 compares the characteristics of ED hospitalizations in which an unplanned transfer occurred to those that did not experience an unplanned transfer. Unplanned transfers were more likely to have a higher comorbidity burden, more deranged physiology, and more likely to arrive on the floor during the overnight shift.
Characteristics | Unplanned Transfer to ICU Within 24 h of Leaving ED? | P Value* | |
---|---|---|---|
Yes | No | ||
N = 4,252 (2.4%) | N = 174,063 (97.6%) | ||
| |||
Age, median (IQR) | 69 (5680) | 70 (5681) | <0.01 |
Male, % | 51.3 | 45.9 | <0.01 |
Comorbidity Points Score (COPS), median (IQR) | 100 (46158) | 89 (42144) | <0.01 |
Laboratory Acute Physiology Score (LAPS), median (IQR) | 26 (1342) | 18 (633) | <0.01 |
Nursing shift on arrival to floor, % | |||
Day: 7 am3 pm (Reference) | 20.1 | 20.1 | NS |
Evening: 3 pm11 pm | 47.6 | 50.2 | NS |
Overnight: 11 pm7 am | 32.3 | 29.7 | <0.01 |
Weekend admission, % | 33.7 | 32.7 | NS |
Admitted to monitored bed, % | 24.1 | 24.9 | NS |
Emergency department annual volume, mean (SD) | 48,755 (15,379) | 50,570 (15,276) | <0.01 |
Non‐ICU annual admission volume, mean (SD) | 5,562 (1,626) | 5,774 (1,568) | <0.01 |
Admitting diagnosis, listed by descending frequency, % | NS | ||
Pneumonia and respiratory infections | 16.3 | 11.8 | <0.01 |
Gastrointestinal bleeding | 12.8 | 13.6 | NS |
Chest pain | 7.3 | 10.0 | <0.01 |
Miscellaneous conditions | 5.6 | 6.2 | NS |
All other acute infections | 4.7 | 6.0 | <0.01 |
Seizures | 4.1 | 5.9 | <0.01 |
AMI | 3.9 | 3.3 | <0.05 |
COPD | 3.8 | 3.0 | <0.01 |
CHF | 3.5 | 3.7 | NS |
Arrhythmias and pulmonary embolism | 3.5 | 3.3 | NS |
Stroke | 3.4 | 3.5 | NS |
Diabetic emergencies | 3.3 | 2.6 | <0.01 |
Metabolic, endocrine, electrolytes | 3.0 | 2.9 | NS |
Sepsis | 3.0 | 1.2 | <0.01 |
Other neurology and toxicology | 3.0 | 2.9 | NS |
Urinary tract infections | 2.9 | 3.2 | NS |
Catastrophic conditions | 2.6 | 1.2 | <0.01 |
Rheumatology | 2.5 | 3.5 | <0.01 |
Hematology and oncology | 2.4 | 2.4 | NS |
Acute renal failure | 1.9 | 1.1 | <0.01 |
Pancreatic and liver | 1.7 | 2.0 | NS |
Trauma, fractures, and dislocations | 1.6 | 1.8 | NS |
Bowel obstructions and diseases | 1.6 | 2.9 | <0.01 |
Other cardiac conditions | 1.5 | 1.3 | NS |
Other renal conditions | 0.6 | 1.0 | <0.01 |
Inpatient length of stay, median days (IQR) | 4.7 (2.78.6) | 2.6 (1.54.4) | <0.01 |
Died during hospitalization, % | 12.7 | 2.4 | <0.01 |
Unplanned ICU transfers were more frequent in lower volume hospitals (Table 1). Figure 1 displays the inverse relationship between hospital annual ED inpatient admission volume and unplanned ICU transfers rates. The lowest volume hospital had a crude rate twice as high as the 2 highest volume hospitals (39 vs 20, per 1,000 admissions).
Pneumonia/respiratory infection was the most frequent admitting condition associated with unplanned transfer (16.3%) (Table 1). There was also wide variation in crude rates for unplanned ICU transfer by admitting condition (Figure 2). Patients admitted with sepsis had the highest rate (59 per 1,000 admissions), while patients admitted with renal conditions other than acute renal failure had the lowest rates (14.3 per 1,000 admissions).
We confirmed that almost all diagnoses found to account for a disproportionately high share of unplanned ICU transfers in Table 1 were indeed independently associated with this phenomenon after adjustment for patient and hospital differences (Figure 2). Pneumonia remained the most frequent condition associated with unplanned ICU transfer (odds ratio [OR] 1.50; 95% CI 1.201.86). Although less frequent, sepsis had the strongest association of any condition with unplanned transfer (OR 2.51; 95% CI 1.903.31). However, metabolic, endocrine, and electrolyte conditions were no longer associated with unplanned transfer after adjustment, while arrhythmias and pulmonary embolism were. Other conditions confirmed to be associated with increased risk of unplanned transfer included: myocardial infarction (MI), chronic obstructive pulmonary disease (COPD), stroke, diabetic emergencies, catastrophic conditions (includes aortic catastrophes, all forms of shock except septic shock, and intracranial hemorrhage), and acute renal failure. After taking into account the frequency of admitting diagnoses, respiratory conditions (COPD, pneumonia/acute respiratory infection) comprised nearly half (47%) of all conditions associated with increased risk of unplanned ICU transfer.
Other factors confirmed to be independently associated with unplanned ICU transfer included: male sex (OR 1.20; 95% CI 1.131.28), high comorbidity burden as measured by COPS >145 (OR 1.13; 95% CI 1.031.24), increasingly abnormal physiology compared to a LAPS <7, and arrival on ward during the overnight shift (OR 1.10; 95% CI 1.011.21). After adjustment, we did find that admission to the TCU rather than a medicalsurgical unit was associated with decreased risk of unplanned ICU transfer (OR 0.83; 95% CI 0.770.90). Age 85 was associated with decreased risk of unplanned ICU transfer relative to the youngest age group of 1834‐year‐old patients (OR 0.64; 95% CI 0.530.77).
ED admissions to higher volume hospitals were 6% less likely to experience an unplanned transfer for each additional 1,000 annual ED hospitalizations over a lower volume hospital (OR 0.94; 95% CI 0.910.98). In other words, a patient admitted to a hospital with 8,000 annual ED hospitalizations had 30% decreased odds of unplanned ICU transfer compared to a hospital with only 3,000 annual ED hospitalizations.
DISCUSSION
Patients admitted with respiratory conditions accounted for half of all admitting diagnoses associated with increased risk of unplanned transfer to the ICU within 24 hours of arrival to the ward. We found that 1 in 30 ED ward admissions for pneumonia, and 1 in 33 for COPD, were transferred to the ICU within 24 hours. These findings indicate that there is some room for improvement in early care of respiratory conditions, given the average unplanned transfer rate of 1 in 42, and previous research showing that patients with pneumonia and patients with COPD, who experience unplanned ICU transfer, have substantially worse mortality than those directly admitted to the ICU.1
Although less frequent than hospitalizations for respiratory conditions, patients admitted with sepsis were at the highest risk of unplanned ICU transfer (1 in 17 ED non‐ICU hospitalizations). We also found that MI and stroke ward admissions had a higher risk of unplanned ICU transfer. However, we previously found that unplanned ICU transfers for sepsis, MI, and stroke did not have worse mortality than direct ICU admits for these conditions.1 Therefore, quality improvement efforts to reduce excess mortality related to early decompensation in the hospital and unplanned ICU transfer would be most effective if targeted towards respiratory conditions such as pneumonia and COPD.
This is the only in‐depth study, to our knowledge, to explore the association between a set of mutually exclusive diagnostic categories and risk of unplanned ICU transfer within 24 hours, and it is the first study to identify risk factors for unplanned ICU transfer in a multi‐hospital cohort adjusted for patient‐ and hospital‐level characteristics. We also identified a novel hospital volumeoutcome relationship: Unplanned ICU transfers are up to twice as likely to occur in the smallest volume hospitals compared with highest volume hospitals. Hospital volume has long been proposed as a proxy for hospital resources; there are several studies showing a relationship between low‐volume hospitals and worse outcomes for a number of conditions.19, 20 Possible mechanisms may include decreased ICU capacity, decreased on‐call intensivists in the hospital after hours, and less experience with certain critical care conditions seen more frequently in high‐volume hospitals.21
Patients at risk of unplanned ICU transfer were also more likely to have physiologic derangement identified on laboratory testing, high comorbidity burden, and arrive on the ward between 11 PM and 7 AM. Given the strong correlation between comorbidity burden and physiologic derangement and mortality,14 it is not surprising that the COPS and LAPS were independent predictors of unplanned transfer. It is unclear, however, why arriving on the ward on the overnight shift is associated with higher risk. One possibility is that patients who arrive on the wards during 11 PM to 7 AM are also likely to have been in the ED during evening peak hours most associated with ED crowding.22 High levels of ED crowding have been associated with delays in care, worse quality care, lapses in patient safety, and even increased in‐hospital mortality.22, 23 Other possible reasons include decreased in‐hospital staffing and longer delays in critical diagnostic tests and interventions.2428
Admission to TCUs was associated with decreased risk of unplanned ICU transfer in the first 24 hours of hospitalization. This may be due to the continuous monitoring, decreased nursing‐to‐patient ratios, or the availability to provide some critical care interventions. In our study, age 85 was associated with lower likelihood of unplanned transfer. Unfortunately, we did not have access to data on advanced directives or patient preferences. Data on advanced directives would help to distinguish whether this phenomenon was related to end‐of‐life care goals versus other explanations.
Our study confirms some risk factors identified in previous studies. These include specific diagnoses such as pneumonia and COPD,12, 13, 29 heavy comorbidity burden,12, 13, 29 abnormal labs,29 and male sex.13 Pneumonia has consistently been shown to be a risk factor for unplanned ICU transfer. This may stem from the dynamic nature of this condition and its ability to rapidly progress, and the fact that some ICUs may not accept pneumonia patients unless they demonstrate a need for mechanical ventilation.30 Recently, a prediction rule has been developed to determine which patients with pneumonia are likely to have an unplanned ICU transfer.30 It is possible that with validation and application of this rule, unplanned transfer rates for pneumonia could be reduced. It is unclear whether males have unmeasured factors associated with increased risk of unplanned transfer or whether a true gender disparity exists.
Our findings should be interpreted within the context of this study's limitations. First, this study was not designed to distinguish the underlying cause of the unplanned transfer such as under‐recognition of illness severity in the ED, evolving clinical disease after leaving the ED, or delays in critical interventions on the ward. These are a focus of our ongoing research efforts. Second, while previous studies have demonstrated that our automated risk adjustment variables can accurately predict in‐hospital mortality (0.88 area under curve in external populations),17 additional data on vital signs and mental status could further improve risk adjustment. However, using automated data allowed us to study risk factors for unplanned transfer in a multi‐hospital cohort with a much larger population than has been previously studied. Serial data on vital signs and mental status both in the ED and during hospitalization could also be helpful in determining which unplanned transfers could be prevented with earlier recognition and intervention. Finally, all patient care occurred within an integrated healthcare delivery system. Thus, differences in case‐mix, hospital resources, ICU structure, and geographic location should be considered when applying our results to other healthcare systems.
This study raises several new areas for future research. With access to richer data becoming available in electronic medical records, prediction rules should be developed to enable better triage to appropriate levels of care for ED admissions. Future research should also analyze the comparative effectiveness of intermediate monitored units versus non‐monitored wards for preventing clinical deterioration by admitting diagnosis. Diagnoses that have been shown to have an increased risk of death after unplanned ICU transfer, such as pneumonia/respiratory infection and COPD,1 should be prioritized in this research. Better understanding is needed on the diagnosis‐specific differences and the differences in ED triage process and ICU structure that may explain why high‐volume hospitals have significantly lower rates of early unplanned ICU transfers compared with low‐volume hospitals. In particular, determining the effect of TCU and ICU capacities and census at the time of admission, and comparing patient risk characteristics across hospital‐volume strata would be very useful. Finally, more work is needed to determine whether the higher rate of unplanned transfers during overnight nursing shifts is related to decreased resource availability, preceding ED crowding, or other organizational causes.
In conclusion, patients admitted with respiratory conditions, sepsis, MI, high comorbidity, and abnormal labs are at modestly increased risk of unplanned ICU transfer within 24 hours of admission from the ED. Patients admitted with respiratory conditions (pneumonia/respiratory infections and COPD) accounted for half of the admitting diagnoses that are at increased risk for unplanned ICU transfer. These patients may benefit from better inpatient triage from the ED, earlier intervention, or closer monitoring. More research is needed to determine the specific aspects of care associated with admission to intermediate care units and high‐volume hospitals that reduce the risk of unplanned ICU transfer.
Acknowledgements
The authors thank John D. Greene, Juan Carlos La Guardia, and Benjamin Turk for their assistance with formatting of the dataset; Dr Alan S. Go, Acting Director of the Division of Research, for reviewing the manuscript; and Alina Schnake‐Mahl for formatting the manuscript.
- Adverse outcomes associated with delayed intensive care unit transfers in an integrated healthcare system. J Hosp Med. 2011;7(3):224–230. , , , et al.
- Inpatient transfers to the intensive care unit. J Gen Intern Med. 2003;18(2):77–83. , , , et al.
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6:74–80. , , , et al.
- Hospital‐wide code rates and mortality before and after implementation of a rapid response team. JAMA. 2008;300(21):2506–2513. , , , et al.
- Effect of a rapid response team on hospital‐wide mortality and code rates outside the ICU in a children's hospital. JAMA. 2007;298(19):2267–2274. , , , et al.
- Introduction of the medical emergency team (MET) system: a cluster‐randomised controlled trial. Lancet. 2005;365(9477):2091–2097. , , , et al.
- Rapid response systems: A systematic review. Crit Care Med. 2007;35(5):1238–1243. , , , et al.
- Effects of rapid response systems on clinical outcomes: systematic review and meta‐analysis. J Hosp Med. 2007;2(6):422–432. , , , et al.
- Rapid response teams: a systematic review and meta‐analysis. Arch Intern Med. 2010;170(1):18–26. , , , et al.
- Outreach and early warning systems (EWS) for the prevention of intensive care admission and death of critically ill adult patients on general hospital wards. Cochrane Database Syst Rev. 2007;3:CD005529. , , , et al.
- Unplanned transfers to a medical intensive care unit: Causes and relationship to preventable errors in care. J Hosp Med. 2011;6:68–72. , , , et al.
- Using administrative data to develop a nomogram for individualising risk of unplanned admission to intensive care. Resuscitation. 2008;79(2):241–248. , , , et al.
- Unplanned admission to intensive care after emergency hospitalisation: risk factors and development of a nomogram for individualising risk. Resuscitation. 2009;80(2):224–230. , , , et al.
- Risk‐adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , et al.
- Linking automated databases for research in managed care settings. Ann Intern Med. 1997;127(8 pt 2):719–724. .
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases. Am J Manag Care. 2008;14(3):158–166. , , , et al.
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population. J Clin Epidemiol. 2011;63(7):798–803. , , , et al.
- Maximum likelihood estimation of limited and discrete dependent variable models with nested random effects. J Econometrics. 2005;128(2):301–323. , , .
- The relation between volume and outcome in health care. N Engl J Med. 1999;340(21):1677–1679. .
- Is volume related to outcome in health care? A systematic review and methodologic critique of the literature. Ann Intern Med. 2002;137(6):511–520. , , .
- Working with capacity limitations: operations management in critical care. Crit Care. 2011;15(4):308. , , .
- Systematic review of emergency department crowding: causes, effects, and solutions. Ann Intern Med. 2008;52(2):126–136. , .
- The effect of emergency department crowding on clinically oriented outcomes. Acad Emerg Med. 2009;16(1):1–10. , , , et al.
- Association between time of admission to the ICU and mortality. Chest. 2010;138(1):68–75. , , , et al.
- Off‐hour admission and in‐hospital stroke case fatality in the get with the guidelines‐stroke program. Stroke. 2009;40(2):569–576. , , , et al.
- Relationship between time of day, day of week, timeliness of reperfusion, and in‐hospital mortality for patients with acute ST‐segment elevation myocardial infarction. JAMA. 2005;294(7):803–812. , , , et al.
- Hospital mortality among adults admitted to and discharged from intensive care on weekends and evenings. J Crit Care. 2008;23(3):317–324. , , , et al.
- Association between ICU admission during morning rounds and mortality. Chest. 2009;136(6):1489–1495. , , , et al.
- Identifying infected emergency department patients admitted to the hospital ward at risk of clinical deterioration and intensive care unit transfer. Acad Emerg Med. 2010;17(10):1080–1085. , , , et al.
- Risk stratification of early admission to the intensive care unit of patients with no major criteria of severe community‐acquired pneumonia: development of an international prediction rule. Crit Care. 2009;13(2):R54. , , , et al.
- Adverse outcomes associated with delayed intensive care unit transfers in an integrated healthcare system. J Hosp Med. 2011;7(3):224–230. , , , et al.
- Inpatient transfers to the intensive care unit. J Gen Intern Med. 2003;18(2):77–83. , , , et al.
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS). J Hosp Med. 2011;6:74–80. , , , et al.
- Hospital‐wide code rates and mortality before and after implementation of a rapid response team. JAMA. 2008;300(21):2506–2513. , , , et al.
- Effect of a rapid response team on hospital‐wide mortality and code rates outside the ICU in a children's hospital. JAMA. 2007;298(19):2267–2274. , , , et al.
- Introduction of the medical emergency team (MET) system: a cluster‐randomised controlled trial. Lancet. 2005;365(9477):2091–2097. , , , et al.
- Rapid response systems: A systematic review. Crit Care Med. 2007;35(5):1238–1243. , , , et al.
- Effects of rapid response systems on clinical outcomes: systematic review and meta‐analysis. J Hosp Med. 2007;2(6):422–432. , , , et al.
- Rapid response teams: a systematic review and meta‐analysis. Arch Intern Med. 2010;170(1):18–26. , , , et al.
- Outreach and early warning systems (EWS) for the prevention of intensive care admission and death of critically ill adult patients on general hospital wards. Cochrane Database Syst Rev. 2007;3:CD005529. , , , et al.
- Unplanned transfers to a medical intensive care unit: Causes and relationship to preventable errors in care. J Hosp Med. 2011;6:68–72. , , , et al.
- Using administrative data to develop a nomogram for individualising risk of unplanned admission to intensive care. Resuscitation. 2008;79(2):241–248. , , , et al.
- Unplanned admission to intensive care after emergency hospitalisation: risk factors and development of a nomogram for individualising risk. Resuscitation. 2009;80(2):224–230. , , , et al.
- Risk‐adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases. Med Care. 2008;46(3):232–239. , , , et al.
- Linking automated databases for research in managed care settings. Ann Intern Med. 1997;127(8 pt 2):719–724. .
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases. Am J Manag Care. 2008;14(3):158–166. , , , et al.
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population. J Clin Epidemiol. 2011;63(7):798–803. , , , et al.
- Maximum likelihood estimation of limited and discrete dependent variable models with nested random effects. J Econometrics. 2005;128(2):301–323. , , .
- The relation between volume and outcome in health care. N Engl J Med. 1999;340(21):1677–1679. .
- Is volume related to outcome in health care? A systematic review and methodologic critique of the literature. Ann Intern Med. 2002;137(6):511–520. , , .
- Working with capacity limitations: operations management in critical care. Crit Care. 2011;15(4):308. , , .
- Systematic review of emergency department crowding: causes, effects, and solutions. Ann Intern Med. 2008;52(2):126–136. , .
- The effect of emergency department crowding on clinically oriented outcomes. Acad Emerg Med. 2009;16(1):1–10. , , , et al.
- Association between time of admission to the ICU and mortality. Chest. 2010;138(1):68–75. , , , et al.
- Off‐hour admission and in‐hospital stroke case fatality in the get with the guidelines‐stroke program. Stroke. 2009;40(2):569–576. , , , et al.
- Relationship between time of day, day of week, timeliness of reperfusion, and in‐hospital mortality for patients with acute ST‐segment elevation myocardial infarction. JAMA. 2005;294(7):803–812. , , , et al.
- Hospital mortality among adults admitted to and discharged from intensive care on weekends and evenings. J Crit Care. 2008;23(3):317–324. , , , et al.
- Association between ICU admission during morning rounds and mortality. Chest. 2009;136(6):1489–1495. , , , et al.
- Identifying infected emergency department patients admitted to the hospital ward at risk of clinical deterioration and intensive care unit transfer. Acad Emerg Med. 2010;17(10):1080–1085. , , , et al.
- Risk stratification of early admission to the intensive care unit of patients with no major criteria of severe community‐acquired pneumonia: development of an international prediction rule. Crit Care. 2009;13(2):R54. , , , et al.
Copyright © 2012 Society of Hospital Medicine
Detection of Physiologic Deterioration
Patients in general medicalsurgical wards who experience unplanned transfer to the intensive care unit (ICU) have increased mortality and morbidity.13 Using an externally validated methodology permitting assessment of illness severity and mortality risk among all hospitalized patients,4, 5 we recently documented observed‐to‐expected mortality ratios >3.0 and excess length of stay of 10 days among patients who experienced such transfers.6
It is possible to predict adverse outcomes among monitored patients (eg, patients in the ICU or undergoing continuous electronic monitoring).7, 8 However, prediction of unplanned transfers among medicalsurgical ward patients presents challenges. Data collection (vital signs and laboratory tests) is relatively infrequent. The event rate (3% of hospital admissions) is low, and the rate in narrow time periods (eg, 12 hours) is extremely low: a hospital with 4000 admissions per year might experience 1 unplanned transfer to the ICU every 3 days. Not surprisingly, performance of models suitable for predicting ward patients' need for intensive care within narrow time frames have been disappointing.9 The Modified Early Warning Score (MEWS), has a c‐statistic, or area under the receiver operator characteristic of 0.67,1012 and our own model incorporating 14 laboratory tests, but no vital signs, has excellent performance with respect to predicting inpatient mortality, but poor performance with respect to unplanned transfer.6
In this report, we describe the development and validation of a complex predictive model suitable for use with ward patients. Our objective for this work was to develop a predictive model based on clinical and physiologic data available in real time from a comprehensive electronic medical record (EMR), not a clinically intuitive, manually assigned tool. The outcome of interest was unplanned transfer from the ward to the ICU, or death on the ward in a patient who was full code. This model has been developed as part of a regional effort to decrease preventable mortality in the Northern California Kaiser Permanente Medical Care Program (KPMCP), an integrated healthcare delivery system with 22 hospitals.
MATERIALS AND METHODS
For additional details, see the Supporting Information, Appendices 112, in the online version of this article.
This project was approved by the KPMCP Institutional Board for the Protection of Human Subjects.
The Northern California KPMCP serves a total population of approximately 3.3 million members. All Northern California KPMCP hospitals and clinics employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere. Databases maintained by the KPMCP capture admission and discharge times, admission and discharge diagnoses and procedures (assigned by professional coders), bed histories permitting quantification of intra‐hospital transfers, inter‐hospital transfers, as well as the results of all inpatient and outpatient laboratory tests. In July 2006, the KPMCP began deployment of the EMR developed by Epic Systems Corporation (
Our setting consisted of 14 hospitals in which the KPHC inpatient EMR had been running for at least 3 months (the KPMCP Antioch, Fremont, Hayward, Manteca, Modesto, Roseville, Sacramento, Santa Clara, San Francisco, Santa Rosa, South Sacramento, South San Francisco, Santa Teresa, and Walnut Creek hospitals). We have described the general characteristics of KPMCP hospitals elsewhere.4, 6 Our initial study population consisted of all patients admitted to these hospitals who met the following criteria: hospitalization began from November 1, 2006 through December 31, 2009; initial hospitalization occurred at a Northern California KPMCP hospital (ie, for inter‐hospital transfers, the first hospital stay occurred within the KPMCP); age 18 years; hospitalization was not for childbirth; and KPHC had been operational at the hospital for at least 3 months.
Analytic Approach
The primary outcome for this study was transfer to the ICU after admission to the hospital among patients residing either in a general medicalsurgical ward (ward) or transitional care unit (TCU), or death in the ward or TCU in a patient who was full code at the time of death (ie, had the patient survived, s/he would have been transferred to the ICU). The unit of analysis for this study was a 12‐hour patient shift, which could begin with a 7 AM T0 (henceforth, day shift) or a 7 PM T0 (night shift); in other words, we aimed to predict the occurrence of an event within 12 hours of T0 using only data available prior to T0. A shift in which a patient experienced the primary study outcome is an event shift, while one in which a patient did not experience the primary outcome is a comparison shift. Using this approach, an individual patient record could consist of both event and comparison shifts, since some patients might have multiple unplanned transfers and some patients might have none. Our basic analytic approach consisted of creating a cohort of event and comparison shifts (10 comparison shifts were randomly selected for each event shift), splitting the cohort into a derivation dataset (50%) and validation dataset (50%), developing a model using the derivation dataset, then applying the coefficients of the derivation dataset to the validation dataset. Because some event shifts were excluded due to the minimum 4‐hour length‐of‐stay requirement, we also applied model coefficients to these excluded shifts and a set of randomly selected comparison shifts.
Since the purpose of these analyses was to develop models with maximal signal extraction from sparsely collected predictors, we did not block a time period after the T0 to allow for a reaction time to the alarm. Thus, since some events could occur immediately after the T0 (as can be seen in the Supporting Information, Appendices, in the online version of this article), our models would need to be run at intervals that are more frequent than 2 times a day.
Independent Variables
In addition to patients' age and sex, we tested the following candidate independent variables. Some of these variables are part of the KPMCP risk adjustment model4, 5 and were available electronically for all patients in the cohort. We grouped admission diagnoses into 44 broad diagnostic categories (primary conditions), and admission types into 4 groups (emergency medical, emergency surgical, elective medical, and elective surgical). We quantified patients' degree of physiologic derangement in the 72 hours preceding hospitalization with a Laboratory‐based Acute Physiology Score (LAPS) using 14 laboratory test results prior to hospitalization; we also tested individual laboratory test results obtained after admission to the hospital. We quantified patients' comorbid illness burden using a COmorbidity Point Score (COPS) based on patients' preexisting diagnoses over the 12‐month period preceding hospitalization.4 We extracted temperature, heart rate, respiratory rate, systolic blood pressure, diastolic blood pressure, oxygen saturation, and neurological status from the EMR. We also tested the following variables based on specific information extracted from the EMR: shock index (heart rate divided by systolic blood pressure)13; care directive status (patients were placed into 4 groups: full code, partial code, do not resuscitate [DNR], and no care directive in place); and a proxy for measured lactate (PML; anion gap/serum bicarbonate 100).1416 For comparison purposes, we also created a retrospective electronically assigned MEWS, which we refer to as the MEWS(re), and we assigned this score to patient records electronically using data from KP HealthConnect.
Statistical Methods
Analyses were performed in SAS 9.1, Stata 10, and R 2.12. Final validation was performed using SAS (SAS Institute Inc., Carey, North Carolina). Since we did not limit ourselves to traditional severity‐scoring approaches (eg, selecting the worst heart rate in a given time interval), but also included trend terms (eg, change in heart rate over the 24 hours preceding T0), the number of potential variables to test was very large. Detailed description of the statistical strategies employed for variable selection is provided in the Supporting Information, Appendices, in the online version of this article. Once variables were selected, our basic approach was to test a series of diagnosis‐specific logistic regression submodels using a variety of predictors that included vital signs, vital signs trends (eg, most recent heart rate minus earliest heart rate, heart rate over preceding 24 hours), and other above‐mentioned variables.
We assessed the ability of a submodel to correctly distinguish patients who died, from survivors, using the c‐statistic, as well as other metrics recommended by Cook.17 At the end of the modeling process, we pooled the results across all submodels. For vital signs, where the rate of missing data was <3%, we tested submodels in which we dropped shifts with missing data, as well as submodels in which we imputed missing vital signs to a normal value. For laboratory data, where the rate of missing data for a given shift was much greater, we employed a probabilistic imputation method that included consideration of when a laboratory test result became available.
RESULTS
During the study period, a total of 102,488 patients experienced 145,335 hospitalizations at the study hospitals. We removed 66 patients with 138 hospitalizations for data quality reasons, leaving us with our initial study sample of 102,422 patients whose characteristics are summarized in Table 1. Table 1, in which the unit of analysis is an individual patient, shows that patients who experienced the primary outcome were similar to those patients described in our previous report, in terms of their characteristics on admission as well as in experiencing excess morbidity and mortality.6
Never Admitted to ICU | Direct Admit to ICU From ED | Unplanned Transfer to ICU* | Other ICU Admission | |
---|---|---|---|---|
| ||||
N | 89,269 | 5963 | 2880 | 4310 |
Age (mean SD) | 61.26 18.62 | 62.25 18.13 | 66.12 16.20 | 64.45 15.91 |
Male (n, %) | 37,228 (41.70%) | 3091 (51.84%) | 1416 (49.17%) | 2378 (55.17%) |
LAPS (mean SD) | 13.02 15.79 | 32.72 24.85 | 24.83 21.53 | 11.79 18.16 |
COPS(mean SD) | 67.25 51.42 | 73.88 57.42 | 86.33 59.33 | 78.44 52.49 |
% Predicted mortality risk (mean SD) | 1.93% 3.98% | 7.69% 12.59% | 5.23% 7.70% | 3.66% 6.81% |
Survived first hospitalization to discharge∥ | 88,479 (99.12%) | 5336 (89.49%) | 2316 (80.42%) | 4063 (94.27%) |
Care order on admission | ||||
Full code | 78,877 (88.36%) | 5198 (87.17%) | 2598 (90.21%) | 4097 (95.06%) |
Partial code | 664 (0.74%) | 156 (2.62%) | 50 (1.74%) | 27 (0.63%) |
Comfort care | 21 (0.02%) | 2 (0.03%) | 0 (0%) | 0 (0%) |
DNR | 8227 (9.22%) | 539 (9.04%) | 219 (7.60%) | 161 (3.74%) |
Comfort care and DNR | 229 (0.26%) | 9 (0.15%) | 2 (0.07%) | 2 (0.05%) |
No order | 1251 (1.40%) | 59 (0.99%) | 11 (0.38%) | 23 (0.53%) |
Admission diagnosis (n, %) | ||||
Pneumonia | 2385 (2.67%) | 258 (4.33%) | 242 (8.40%) | 68 (1.58%) |
Sepsis | 5822 (6.52%) | 503 (8.44%) | 279 (9.69%) | 169 (3.92%) |
GI bleeding | 9938 (11.13%) | 616 (10.33%) | 333 (11.56%) | 290 (6.73%) |
Cancer | 2845 (3.19%) | 14 (0.23%) | 95 (3.30%) | 492 (11.42%) |
Total hospital length of stay (days SD) | 3.08 3.29 | 5.37 7.50 | 12.16 13.12 | 8.06 9.53 |
Figure 1shows how we developed the analysis cohort, by removing patients with a comfort‐care‐only order placed within 4 hours after admission (369 patients/744 hospitalizations) and patients who were never admitted to the ward or TCU (7,220/10,574). This left a cohort consisting of 94,833 patients who experienced 133,879 hospitalizations spanning a total of 1,079,062 shifts. We then removed shifts where: 1) a patient was not on the ward at the start of a shift, or was on the ward for <4 hours of a shift; 2) the patient had a comfort‐care order in place at the start of the shift; and 3) the patient died and was ineligible to be a case (the patient had a DNR order in place or died in the ICU). The final cohort eligible for sampling consisted of 846,907 shifts, which involved a total of 92,797 patients and 130,627 hospitalizations. There were a total of 4,036 event shifts, which included 3,224 where a patient was transferred from the ward to the ICU, 717 from the TCU to the ICU, and 95 where a patient died on the ward or TCU without a DNR order in place. We then randomly selected 39,782 comparison shifts. Thus, our final cohort for analysis included 4,036 event shifts (1,979 derivation/2,057 validation and 39,782 comparison shifts (19,509/20,273). As a secondary validation, we also applied model coefficients to the 429 event shifts excluded due to the <4‐hour length‐of‐stay requirement.
Table 2 compares event shifts with comparison shifts. In the 24 hours preceding ICU transfer, patients who were subsequently transferred had statistically significant, but not necessarily clinically significant, differences in terms of these variables. However, missing laboratory data were more common, ranging from 18% to 31% of all shifts (we did not incorporate laboratory tests where 35% of the shifts had missing data for that test).
Predictor | Event Shifts | Comparison Shifts | P |
---|---|---|---|
| |||
Number | 4036 | 39,782 | |
Age (mean SD) | 67.19 15.25 | 65.41 17.40 | <0.001 |
Male (n, %) | 2007 (49.73%) | 17,709 (44.52%) | <0.001 |
Day shift | 1364 (33.80%) | 17,714 (44.53%) | <0.001 |
LAPS* | 27.89 22.10 | 20.49 20.16 | <0.001 |
COPS | 116.33 72.31 | 100.81 68.44 | <0.001 |
Full code (n, %) | 3496 (86.2%) | 32,156 (80.8%) | <0.001 |
ICU shift during hospitalization | 3964 (98.22%) | 7197 (18.09%) | <0.001 |
Unplanned transfer to ICU during hospitalization∥ | 353 (8.8%) | 1466 (3.7%) | <0.001 |
Temperature (mean SD) | 98.15 (1.13) | 98.10 (0.85) | 0.009 |
Heart rate (mean SD) | 90.30 (20.48) | 79.86 (5.27) | <0.001 |
Respiratory rate (mean SD) | 20.36 (3.70) | 18.87 (1.79) | <0.001 |
Systolic blood pressure (mean SD) | 123.65 (23.26) | 126.21 (19.88) | <0.001 |
Diastolic blood pressure (mean SD) | 68.38 (14.49) | 69.46 (11.95) | <0.001 |
Oxygen saturation (mean SD) | 95.72% (3.00) | 96.47 % (2.26) | <0.001 |
MEWS(re) (mean SD) | 3.64 (2.02) | 2.34 (1.61) | <0.001 |
% <5 | 74.86% | 92.79% | |
% 5 | 25.14% | 7.21% | <0.001 |
Proxy for measured lactate# (mean SD) | 36.85 (28.24) | 28.73 (16.74) | <0.001 |
% Missing in 24 hr before start of shift** | 17.91% | 28.78% | <0.001 |
Blood urea nitrogen (mean SD) | 32.03 (25.39) | 22.72 (18.9) | <0.001 |
% Missing in 24 hr before start of shift | 19.67% | 20.90% | <0.001 |
White blood cell count 1000 (mean SD) | 12.33 (11.42) | 9.83 (6.58) | <0.001 |
% Missing in 24 hr before start of shift | 21.43% | 30.98% | <0.001 |
Hematocrit (mean SD) | 33.08 (6.28) | 33.07 (5.25) | 0.978 |
% Missing in 24 hr before start of shift | 19.87% | 29.55% | <0.001 |
After conducting multiple analyses using the derivation dataset, we developed 24 submodels, a compromise between our finding that primary‐condition‐specific models showed better performance and the fact that we had very few events among patients with certain primary conditions (eg, pericarditis/valvular heart disease), which forced us to create composite categories (eg, a category pooling patients with pericarditis, atherosclerosis, and peripheral vascular disease). Table 3 lists variables included in our final submodels.
Variable | Description |
---|---|
| |
Directive status | Full code or not full code |
LAPS* | Admission physiologic severity of illness score (continuous variable ranging from 0 to 256). Standardized and included as LAPS and LAPS squared |
COPS | Comorbidity burden score (continuous variable ranging from 0 to 701). Standardized and included as COPS and COPS squared. |
COPS status | Indicator for absent comorbidity data |
LOS at T0 | Length of stay in the hospital (total time in hours) at the T0; standardized. |
T0 time of day | 7 AM or 7 PM |
Temperature | Worst (highest) temperature in 24 hr preceding T0; variability in temperature in 24 hr preceding T0. |
Heart rate | Most recent heart rate in 24 hr preceding T0; variability in heart rate in 24 hr preceding T0. |
Respiratory rate | Most recent respiratory rate in 24 hr preceding T0; worst (highest) respiratory rate in 24 hr preceding T0; variability in respiratory rate in 24 hr preceding T0. |
Diastolic blood pressure | Most recent diastolic blood pressure in 24 hr preceding T0 transformed by subtracting 70 from the actual value and squaring the result. Any value above 2000 is subsequently then set to 2000, yielding a continuous variable ranging from 0 to 2000. |
Systolic pressure | Variability in systolic blood pressure in 24 hr preceding T0. |
Pulse oximetry | Worst (lowest) oxygen saturation in 24 hr preceding T0; variability in oxygen saturation in 24 hr preceding T0. |
Neurological status | Most recent neurological status check in 24 hr preceding T0. |
Laboratory tests | Blood urea nitrogen |
Proxy for measured lactate = (anion gap serum bicarbonate) 100 | |
Hematocrit | |
Total white blood cell count |
Table 4 summarizes key results in the validation dataset. Across all diagnoses, the MEWS(re) had c‐statistic of 0.709 (95% confidence interval, 0.6970.721) in the derivation dataset and 0.698 (0.6860.710) in the validation dataset. In the validation dataset, the MEWS(re) performed best among patients with a set of gastrointestinal diagnoses (c = 0.792; 0.7260.857) and worst among patients with congestive heart failure (0.541; 0.5000.620). In contrast, across all primary conditions, the EMR‐based models had a c‐statistic of 0.845 (0.8260.863) in the derivation dataset and 0.775 (0.7530.797) in the validation dataset. In the validation dataset, the EMR‐based models also performed best among patients with a set of gastrointestinal diagnoses (0.841; 0.7830.897) and worst among patients with congestive heart failure (0.683; 0.6100.755). A negative correlation (R = 0.63) was evident between the number of event shifts in a submodel and the drop in the c‐statistic seen in the validation dataset.
No. of Shifts in Validation Dataset | c‐Statistic | |||
---|---|---|---|---|
Diagnoses Group* | Event | Comparison | MEWS(re) | EMR Model |
| ||||
Acute myocardial infarction | 36 | 169 | 0.541 | 0.572 |
Diseases of pulmonary circulation and cardiac dysrhythmias | 40 | 329 | 0.565 | 0.645 |
Seizure disorders | 45 | 497 | 0.594 | 0.647 |
Rule out myocardial infarction | 77 | 727 | 0.602 | 0.648 |
Pneumonia | 163 | 847 | 0.741 | 0.801 |
GI diagnoses, set A | 58 | 942 | 0.755 | 0.803 |
GI diagnoses, set B∥ | 256 | 2,610 | 0.772 | 0.806 |
GI diagnoses, set C | 46 | 520 | 0.792 | 0.841 |
All diagnosis | 2,032 | 20,106 | 0.698 | 0.775 |
We also compared model performance when our datasets were restricted to 1 randomly selected observation per patient; in these analyses, the total number of event shifts was 3,647 and the number of comparison shifts was 29,052. The c‐statistic for the MEWS(re) in the derivation dataset was 0.709 (0.6940.725); in the validation dataset, it was 0.698 (0.6920.714). The corresponding values for the EMR‐based models were 0.856 (0.8350.877) and 0.780 (0.7560.804). We also tested models in which, instead of dropping shifts with missing vital signs, we imputed missing vital signs to their normal value. The c‐statistic for the EMR‐based model with imputed vital sign values was 0.842 (0.8230.861) in the derivation dataset and 0.773 (0.7520.794) in the validation dataset. Lastly, we applied model coefficients to a dataset consisting of 4,290 randomly selected comparison shifts plus the 429 shifts excluded because of the 4‐hour length‐of‐stay criterion. The c‐statistic for this analysis was 0.756 (0.7030.809).
As a general rule, the EMR‐based models were more than twice as efficient as the MEWS(re). For example, a MEWS(re) threshold of 6 as the trigger for an alarm would identify 15% of all transfers to the ICU, with 34.4 false alarms for each transfer; in contrast, using the EMR‐based approach to identify 15% of all transfers, there were 14.5 false alarms for each transfer. Applied to the entire KPMCP Northern California Region, using the MEWS(re), a total of 52 patients per day would need to be evaluated, but only 22 per day using the EMR‐based approach. If one employed a MEWS(re) threshold of 4, this would lead to identification of 44% of all transfers, with a ratio of 69 false alarms for each transfer; using the EMR, the ratio would be 34 to 1. Across the entire KPMCP, a total of 276 patients per day (or about 19.5 a day per hospital) would need to be evaluated using the MEWS(re), but only 136 (or about 9.5 per hospital per day) using the EMR.
DISCUSSION
Using data from a large hospital cohort, we have developed a predictive model suitable for use in non‐ICU populations cared for in integrated healthcare settings with fully automated EMRs. The overall performance of our model, which incorporates acute physiology, diagnosis, and longitudinal data, is superior to the predictive ability of a model that can be assigned manually. This is not surprising, given that scoring systems such as the MEWS make an explicit tradeoff losing information found in multiple variables in exchange for ease of manual assignment. Currently, the model described in this report is being implemented in a simulated environment, a final safety test prior to piloting real‐time provision of probability estimates to clinicians and nurses. Though not yet ready for real‐time use, it is reasonable for our model to be tested using the KPHC shadow server, since evaluation in a simulated environment constitutes a critical evaluation step prior to deployment for clinical use. We also anticipate further refinement and revalidation to occur as more inpatient data become available in the KPMCP and elsewhere.
A number of limitations to our approach must be emphasized. In developing our models, we determined that, while modeling by clinical condition was important, the study outcome was rare for some primary conditions. In these diagnostic groups, which accounted for 12.5% of the event shifts and 10.6% of the comparison shifts, the c‐statistic in the validation dataset was <0.70. Since all 22 KPMCP hospitals are now online and will generate an additional 150,000 adult hospitalizations per year, we expect to be able to correct this problem prior to deployment of these models for clinical use. Having additional data will permit us to improve model discrimination and thus decrease the evaluation‐to‐detection ratio. In future iterations of these models, more experimentation with grouping of International Classification of Diseases (ICD) codes may be required. The problem of grouping ICD codes is not an easy one to resolve, in that diagnoses in the grouping must share common pathophysiology while having a grouping with a sufficient number of adverse events for stable statistical models.
Ideally, it would have been desirable to employ a more objective measure of deterioration, since the decision to transfer a patient to the ICU is discretionary. However, we have found that key data points needed to define such a measure (eg, vital signs) are not consistently charted when a patient deterioratesthis is not surprising outside the research setting, given that nurses and physicians involved in a transfer may be focusing on caring for the patient rather than immediately charting. Given the complexities of end‐of‐life‐care decision‐making, we could not employ death as the outcome of interest. A related issue is that our model does not differentiate between reasons for needing transfer to the ICU, an issue recently discussed by Bapoje et al.18
Our model does not address an important issue raised by Bapoje et al18 and Litvak, Pronovost, and others,19, 20 namely, whether a patient should have been admitted to a non‐ICU setting in the first place. Our team is currently developing a model for doing exactly this (providing decision support for triage in the emergency department), but discussion of this methodology is outside the scope of this article.
Because of resource and data limitations, our model also does not include newborns, children, women admitted for childbirth, or patients transferred from non‐KPMCP hospitals. However, the approach described here could serve as a starting point for developing models for these other populations.
The generalizability of our model must also be considered. The Northern California KPMCP is unusual in having large electronic databases that include physiologic as well as longitudinal patient data. Many hospitals cannot take advantage of all the methods described here. However, the methods we employed could be modified for use by hospital systems in countries such as Great Britain and Canada, and entities such as the Veterans Administration Hospital System in the United States. The KPMCP population, an insured population with few barriers to access, is healthier than the general population, and some population subsets are underrepresented in our cohort. Practice patterns may also vary. Nonetheless, the model described here could serve as a good starting point for future collaborative studies, and it would be possible to develop models suitable for use by stand‐alone hospitals (eg, recalibrating so that one used a Charlson comorbidity21 score based on present on‐admission codes rather than the COPS).
The need for early detection of patient deterioration has played a major role in the development of rapid response teams, as well as scores such as the MEWS. In particular, entities such as the Institute for Healthcare Improvement have advocated the use of early warning systems.22 However, having a statistically robust model to support an early warning system is only part of the solution, and a number of new challenges must then be addressed. The first is actual electronic deployment. Existing inpatient EMRs were not designed with complex calculations in mind, and we anticipate that some degradation in performance will occur when we test our models using real‐time data capture. As Bapoje et al point out, simply having an alert may be insufficient, since not all transfers are preventable.18 Early warning systems also raise ethical issues (for example, what should be done if an alert leads a clinician to confront the fact that an end‐of‐life‐care discussion needs to occur?). From a research perspective, if one were to formally test the benefits of such models, it would be critical to define outcome measures other than death (which is strongly affected by end‐of‐life‐care decisions) or ICU transfer (which is often desirable).
In conclusion, we have developed an approach for predicting impending physiologic deterioration of hospitalized adults outside the ICU. Our approach illustrates how organizations can take maximal advantage of EMRs in a manner that exceeds meaningful use specifications.23, 24 Our study highlights the possibility of using fully automated EMR data for building and applying sophisticated statistical models in settings other than the highly monitored ICU without the need for additional equipment. It also expands the universe of severity scoring to one in which probability estimates are provided in real time and throughout an entire hospitalization. Model performance will undoubtedly improve over time, as more patient data become available. Although our approach has important limitations, it is suitable for testing using real‐time data in a simulated environment. Such testing would permit identification of unanticipated problems and quantification of the degradation of model performance due to real life factors, such as delays in vital signs charting or EMR system brownouts. It could also serve as the springboard for future collaborative studies, with a broader population base, in which the EMR becomes a tool for care, not just documentation.
Acknowledgements
We thank Ms Marla Gardner and Mr John Greene for their work in the development phase of this project. We are grateful to Brian Hoberman, Andrew Hwang, and Marc Flagg from the RIMS group; to Colin Stobbs, Sriram Thiruvenkatachari, and Sundeep Sood from KP IT, Inc; and to Dennis Andaya, Linda Gliner, and Cyndi Vasallo for their assistance with data‐quality audits. We are also grateful to Dr Philip Madvig, Dr Paul Feigenbaum, Dr Alan Whippy, Mr Gregory Adams, Ms Barbara Crawford, and Dr Marybeth Sharpe for their administrative support and encouragement; and to Dr Alan S. Go, Acting Director of the Kaiser Permanente Division of Research, for reviewing the manuscript.
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- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS).J Hosp Med.2011;6(2):74–80. , , , , , .
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- Integration of early physiological responses predicts later illness severity in preterm infants.Sci Transl Med.2010;2(48):48ra65. , , , , .
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- MERIT Study Investigators.Introduction of the medical emergency team (MET) system: a cluster‐randomized controlled trial.Lancet.2005;365(9477):2091–2097.
- Unplanned transfers to the intensive care unit: the role of the shock index.J Hosp Med.2010;5(8):460–465. , , , , , .
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Patients in general medicalsurgical wards who experience unplanned transfer to the intensive care unit (ICU) have increased mortality and morbidity.13 Using an externally validated methodology permitting assessment of illness severity and mortality risk among all hospitalized patients,4, 5 we recently documented observed‐to‐expected mortality ratios >3.0 and excess length of stay of 10 days among patients who experienced such transfers.6
It is possible to predict adverse outcomes among monitored patients (eg, patients in the ICU or undergoing continuous electronic monitoring).7, 8 However, prediction of unplanned transfers among medicalsurgical ward patients presents challenges. Data collection (vital signs and laboratory tests) is relatively infrequent. The event rate (3% of hospital admissions) is low, and the rate in narrow time periods (eg, 12 hours) is extremely low: a hospital with 4000 admissions per year might experience 1 unplanned transfer to the ICU every 3 days. Not surprisingly, performance of models suitable for predicting ward patients' need for intensive care within narrow time frames have been disappointing.9 The Modified Early Warning Score (MEWS), has a c‐statistic, or area under the receiver operator characteristic of 0.67,1012 and our own model incorporating 14 laboratory tests, but no vital signs, has excellent performance with respect to predicting inpatient mortality, but poor performance with respect to unplanned transfer.6
In this report, we describe the development and validation of a complex predictive model suitable for use with ward patients. Our objective for this work was to develop a predictive model based on clinical and physiologic data available in real time from a comprehensive electronic medical record (EMR), not a clinically intuitive, manually assigned tool. The outcome of interest was unplanned transfer from the ward to the ICU, or death on the ward in a patient who was full code. This model has been developed as part of a regional effort to decrease preventable mortality in the Northern California Kaiser Permanente Medical Care Program (KPMCP), an integrated healthcare delivery system with 22 hospitals.
MATERIALS AND METHODS
For additional details, see the Supporting Information, Appendices 112, in the online version of this article.
This project was approved by the KPMCP Institutional Board for the Protection of Human Subjects.
The Northern California KPMCP serves a total population of approximately 3.3 million members. All Northern California KPMCP hospitals and clinics employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere. Databases maintained by the KPMCP capture admission and discharge times, admission and discharge diagnoses and procedures (assigned by professional coders), bed histories permitting quantification of intra‐hospital transfers, inter‐hospital transfers, as well as the results of all inpatient and outpatient laboratory tests. In July 2006, the KPMCP began deployment of the EMR developed by Epic Systems Corporation (
Our setting consisted of 14 hospitals in which the KPHC inpatient EMR had been running for at least 3 months (the KPMCP Antioch, Fremont, Hayward, Manteca, Modesto, Roseville, Sacramento, Santa Clara, San Francisco, Santa Rosa, South Sacramento, South San Francisco, Santa Teresa, and Walnut Creek hospitals). We have described the general characteristics of KPMCP hospitals elsewhere.4, 6 Our initial study population consisted of all patients admitted to these hospitals who met the following criteria: hospitalization began from November 1, 2006 through December 31, 2009; initial hospitalization occurred at a Northern California KPMCP hospital (ie, for inter‐hospital transfers, the first hospital stay occurred within the KPMCP); age 18 years; hospitalization was not for childbirth; and KPHC had been operational at the hospital for at least 3 months.
Analytic Approach
The primary outcome for this study was transfer to the ICU after admission to the hospital among patients residing either in a general medicalsurgical ward (ward) or transitional care unit (TCU), or death in the ward or TCU in a patient who was full code at the time of death (ie, had the patient survived, s/he would have been transferred to the ICU). The unit of analysis for this study was a 12‐hour patient shift, which could begin with a 7 AM T0 (henceforth, day shift) or a 7 PM T0 (night shift); in other words, we aimed to predict the occurrence of an event within 12 hours of T0 using only data available prior to T0. A shift in which a patient experienced the primary study outcome is an event shift, while one in which a patient did not experience the primary outcome is a comparison shift. Using this approach, an individual patient record could consist of both event and comparison shifts, since some patients might have multiple unplanned transfers and some patients might have none. Our basic analytic approach consisted of creating a cohort of event and comparison shifts (10 comparison shifts were randomly selected for each event shift), splitting the cohort into a derivation dataset (50%) and validation dataset (50%), developing a model using the derivation dataset, then applying the coefficients of the derivation dataset to the validation dataset. Because some event shifts were excluded due to the minimum 4‐hour length‐of‐stay requirement, we also applied model coefficients to these excluded shifts and a set of randomly selected comparison shifts.
Since the purpose of these analyses was to develop models with maximal signal extraction from sparsely collected predictors, we did not block a time period after the T0 to allow for a reaction time to the alarm. Thus, since some events could occur immediately after the T0 (as can be seen in the Supporting Information, Appendices, in the online version of this article), our models would need to be run at intervals that are more frequent than 2 times a day.
Independent Variables
In addition to patients' age and sex, we tested the following candidate independent variables. Some of these variables are part of the KPMCP risk adjustment model4, 5 and were available electronically for all patients in the cohort. We grouped admission diagnoses into 44 broad diagnostic categories (primary conditions), and admission types into 4 groups (emergency medical, emergency surgical, elective medical, and elective surgical). We quantified patients' degree of physiologic derangement in the 72 hours preceding hospitalization with a Laboratory‐based Acute Physiology Score (LAPS) using 14 laboratory test results prior to hospitalization; we also tested individual laboratory test results obtained after admission to the hospital. We quantified patients' comorbid illness burden using a COmorbidity Point Score (COPS) based on patients' preexisting diagnoses over the 12‐month period preceding hospitalization.4 We extracted temperature, heart rate, respiratory rate, systolic blood pressure, diastolic blood pressure, oxygen saturation, and neurological status from the EMR. We also tested the following variables based on specific information extracted from the EMR: shock index (heart rate divided by systolic blood pressure)13; care directive status (patients were placed into 4 groups: full code, partial code, do not resuscitate [DNR], and no care directive in place); and a proxy for measured lactate (PML; anion gap/serum bicarbonate 100).1416 For comparison purposes, we also created a retrospective electronically assigned MEWS, which we refer to as the MEWS(re), and we assigned this score to patient records electronically using data from KP HealthConnect.
Statistical Methods
Analyses were performed in SAS 9.1, Stata 10, and R 2.12. Final validation was performed using SAS (SAS Institute Inc., Carey, North Carolina). Since we did not limit ourselves to traditional severity‐scoring approaches (eg, selecting the worst heart rate in a given time interval), but also included trend terms (eg, change in heart rate over the 24 hours preceding T0), the number of potential variables to test was very large. Detailed description of the statistical strategies employed for variable selection is provided in the Supporting Information, Appendices, in the online version of this article. Once variables were selected, our basic approach was to test a series of diagnosis‐specific logistic regression submodels using a variety of predictors that included vital signs, vital signs trends (eg, most recent heart rate minus earliest heart rate, heart rate over preceding 24 hours), and other above‐mentioned variables.
We assessed the ability of a submodel to correctly distinguish patients who died, from survivors, using the c‐statistic, as well as other metrics recommended by Cook.17 At the end of the modeling process, we pooled the results across all submodels. For vital signs, where the rate of missing data was <3%, we tested submodels in which we dropped shifts with missing data, as well as submodels in which we imputed missing vital signs to a normal value. For laboratory data, where the rate of missing data for a given shift was much greater, we employed a probabilistic imputation method that included consideration of when a laboratory test result became available.
RESULTS
During the study period, a total of 102,488 patients experienced 145,335 hospitalizations at the study hospitals. We removed 66 patients with 138 hospitalizations for data quality reasons, leaving us with our initial study sample of 102,422 patients whose characteristics are summarized in Table 1. Table 1, in which the unit of analysis is an individual patient, shows that patients who experienced the primary outcome were similar to those patients described in our previous report, in terms of their characteristics on admission as well as in experiencing excess morbidity and mortality.6
Never Admitted to ICU | Direct Admit to ICU From ED | Unplanned Transfer to ICU* | Other ICU Admission | |
---|---|---|---|---|
| ||||
N | 89,269 | 5963 | 2880 | 4310 |
Age (mean SD) | 61.26 18.62 | 62.25 18.13 | 66.12 16.20 | 64.45 15.91 |
Male (n, %) | 37,228 (41.70%) | 3091 (51.84%) | 1416 (49.17%) | 2378 (55.17%) |
LAPS (mean SD) | 13.02 15.79 | 32.72 24.85 | 24.83 21.53 | 11.79 18.16 |
COPS(mean SD) | 67.25 51.42 | 73.88 57.42 | 86.33 59.33 | 78.44 52.49 |
% Predicted mortality risk (mean SD) | 1.93% 3.98% | 7.69% 12.59% | 5.23% 7.70% | 3.66% 6.81% |
Survived first hospitalization to discharge∥ | 88,479 (99.12%) | 5336 (89.49%) | 2316 (80.42%) | 4063 (94.27%) |
Care order on admission | ||||
Full code | 78,877 (88.36%) | 5198 (87.17%) | 2598 (90.21%) | 4097 (95.06%) |
Partial code | 664 (0.74%) | 156 (2.62%) | 50 (1.74%) | 27 (0.63%) |
Comfort care | 21 (0.02%) | 2 (0.03%) | 0 (0%) | 0 (0%) |
DNR | 8227 (9.22%) | 539 (9.04%) | 219 (7.60%) | 161 (3.74%) |
Comfort care and DNR | 229 (0.26%) | 9 (0.15%) | 2 (0.07%) | 2 (0.05%) |
No order | 1251 (1.40%) | 59 (0.99%) | 11 (0.38%) | 23 (0.53%) |
Admission diagnosis (n, %) | ||||
Pneumonia | 2385 (2.67%) | 258 (4.33%) | 242 (8.40%) | 68 (1.58%) |
Sepsis | 5822 (6.52%) | 503 (8.44%) | 279 (9.69%) | 169 (3.92%) |
GI bleeding | 9938 (11.13%) | 616 (10.33%) | 333 (11.56%) | 290 (6.73%) |
Cancer | 2845 (3.19%) | 14 (0.23%) | 95 (3.30%) | 492 (11.42%) |
Total hospital length of stay (days SD) | 3.08 3.29 | 5.37 7.50 | 12.16 13.12 | 8.06 9.53 |
Figure 1shows how we developed the analysis cohort, by removing patients with a comfort‐care‐only order placed within 4 hours after admission (369 patients/744 hospitalizations) and patients who were never admitted to the ward or TCU (7,220/10,574). This left a cohort consisting of 94,833 patients who experienced 133,879 hospitalizations spanning a total of 1,079,062 shifts. We then removed shifts where: 1) a patient was not on the ward at the start of a shift, or was on the ward for <4 hours of a shift; 2) the patient had a comfort‐care order in place at the start of the shift; and 3) the patient died and was ineligible to be a case (the patient had a DNR order in place or died in the ICU). The final cohort eligible for sampling consisted of 846,907 shifts, which involved a total of 92,797 patients and 130,627 hospitalizations. There were a total of 4,036 event shifts, which included 3,224 where a patient was transferred from the ward to the ICU, 717 from the TCU to the ICU, and 95 where a patient died on the ward or TCU without a DNR order in place. We then randomly selected 39,782 comparison shifts. Thus, our final cohort for analysis included 4,036 event shifts (1,979 derivation/2,057 validation and 39,782 comparison shifts (19,509/20,273). As a secondary validation, we also applied model coefficients to the 429 event shifts excluded due to the <4‐hour length‐of‐stay requirement.
Table 2 compares event shifts with comparison shifts. In the 24 hours preceding ICU transfer, patients who were subsequently transferred had statistically significant, but not necessarily clinically significant, differences in terms of these variables. However, missing laboratory data were more common, ranging from 18% to 31% of all shifts (we did not incorporate laboratory tests where 35% of the shifts had missing data for that test).
Predictor | Event Shifts | Comparison Shifts | P |
---|---|---|---|
| |||
Number | 4036 | 39,782 | |
Age (mean SD) | 67.19 15.25 | 65.41 17.40 | <0.001 |
Male (n, %) | 2007 (49.73%) | 17,709 (44.52%) | <0.001 |
Day shift | 1364 (33.80%) | 17,714 (44.53%) | <0.001 |
LAPS* | 27.89 22.10 | 20.49 20.16 | <0.001 |
COPS | 116.33 72.31 | 100.81 68.44 | <0.001 |
Full code (n, %) | 3496 (86.2%) | 32,156 (80.8%) | <0.001 |
ICU shift during hospitalization | 3964 (98.22%) | 7197 (18.09%) | <0.001 |
Unplanned transfer to ICU during hospitalization∥ | 353 (8.8%) | 1466 (3.7%) | <0.001 |
Temperature (mean SD) | 98.15 (1.13) | 98.10 (0.85) | 0.009 |
Heart rate (mean SD) | 90.30 (20.48) | 79.86 (5.27) | <0.001 |
Respiratory rate (mean SD) | 20.36 (3.70) | 18.87 (1.79) | <0.001 |
Systolic blood pressure (mean SD) | 123.65 (23.26) | 126.21 (19.88) | <0.001 |
Diastolic blood pressure (mean SD) | 68.38 (14.49) | 69.46 (11.95) | <0.001 |
Oxygen saturation (mean SD) | 95.72% (3.00) | 96.47 % (2.26) | <0.001 |
MEWS(re) (mean SD) | 3.64 (2.02) | 2.34 (1.61) | <0.001 |
% <5 | 74.86% | 92.79% | |
% 5 | 25.14% | 7.21% | <0.001 |
Proxy for measured lactate# (mean SD) | 36.85 (28.24) | 28.73 (16.74) | <0.001 |
% Missing in 24 hr before start of shift** | 17.91% | 28.78% | <0.001 |
Blood urea nitrogen (mean SD) | 32.03 (25.39) | 22.72 (18.9) | <0.001 |
% Missing in 24 hr before start of shift | 19.67% | 20.90% | <0.001 |
White blood cell count 1000 (mean SD) | 12.33 (11.42) | 9.83 (6.58) | <0.001 |
% Missing in 24 hr before start of shift | 21.43% | 30.98% | <0.001 |
Hematocrit (mean SD) | 33.08 (6.28) | 33.07 (5.25) | 0.978 |
% Missing in 24 hr before start of shift | 19.87% | 29.55% | <0.001 |
After conducting multiple analyses using the derivation dataset, we developed 24 submodels, a compromise between our finding that primary‐condition‐specific models showed better performance and the fact that we had very few events among patients with certain primary conditions (eg, pericarditis/valvular heart disease), which forced us to create composite categories (eg, a category pooling patients with pericarditis, atherosclerosis, and peripheral vascular disease). Table 3 lists variables included in our final submodels.
Variable | Description |
---|---|
| |
Directive status | Full code or not full code |
LAPS* | Admission physiologic severity of illness score (continuous variable ranging from 0 to 256). Standardized and included as LAPS and LAPS squared |
COPS | Comorbidity burden score (continuous variable ranging from 0 to 701). Standardized and included as COPS and COPS squared. |
COPS status | Indicator for absent comorbidity data |
LOS at T0 | Length of stay in the hospital (total time in hours) at the T0; standardized. |
T0 time of day | 7 AM or 7 PM |
Temperature | Worst (highest) temperature in 24 hr preceding T0; variability in temperature in 24 hr preceding T0. |
Heart rate | Most recent heart rate in 24 hr preceding T0; variability in heart rate in 24 hr preceding T0. |
Respiratory rate | Most recent respiratory rate in 24 hr preceding T0; worst (highest) respiratory rate in 24 hr preceding T0; variability in respiratory rate in 24 hr preceding T0. |
Diastolic blood pressure | Most recent diastolic blood pressure in 24 hr preceding T0 transformed by subtracting 70 from the actual value and squaring the result. Any value above 2000 is subsequently then set to 2000, yielding a continuous variable ranging from 0 to 2000. |
Systolic pressure | Variability in systolic blood pressure in 24 hr preceding T0. |
Pulse oximetry | Worst (lowest) oxygen saturation in 24 hr preceding T0; variability in oxygen saturation in 24 hr preceding T0. |
Neurological status | Most recent neurological status check in 24 hr preceding T0. |
Laboratory tests | Blood urea nitrogen |
Proxy for measured lactate = (anion gap serum bicarbonate) 100 | |
Hematocrit | |
Total white blood cell count |
Table 4 summarizes key results in the validation dataset. Across all diagnoses, the MEWS(re) had c‐statistic of 0.709 (95% confidence interval, 0.6970.721) in the derivation dataset and 0.698 (0.6860.710) in the validation dataset. In the validation dataset, the MEWS(re) performed best among patients with a set of gastrointestinal diagnoses (c = 0.792; 0.7260.857) and worst among patients with congestive heart failure (0.541; 0.5000.620). In contrast, across all primary conditions, the EMR‐based models had a c‐statistic of 0.845 (0.8260.863) in the derivation dataset and 0.775 (0.7530.797) in the validation dataset. In the validation dataset, the EMR‐based models also performed best among patients with a set of gastrointestinal diagnoses (0.841; 0.7830.897) and worst among patients with congestive heart failure (0.683; 0.6100.755). A negative correlation (R = 0.63) was evident between the number of event shifts in a submodel and the drop in the c‐statistic seen in the validation dataset.
No. of Shifts in Validation Dataset | c‐Statistic | |||
---|---|---|---|---|
Diagnoses Group* | Event | Comparison | MEWS(re) | EMR Model |
| ||||
Acute myocardial infarction | 36 | 169 | 0.541 | 0.572 |
Diseases of pulmonary circulation and cardiac dysrhythmias | 40 | 329 | 0.565 | 0.645 |
Seizure disorders | 45 | 497 | 0.594 | 0.647 |
Rule out myocardial infarction | 77 | 727 | 0.602 | 0.648 |
Pneumonia | 163 | 847 | 0.741 | 0.801 |
GI diagnoses, set A | 58 | 942 | 0.755 | 0.803 |
GI diagnoses, set B∥ | 256 | 2,610 | 0.772 | 0.806 |
GI diagnoses, set C | 46 | 520 | 0.792 | 0.841 |
All diagnosis | 2,032 | 20,106 | 0.698 | 0.775 |
We also compared model performance when our datasets were restricted to 1 randomly selected observation per patient; in these analyses, the total number of event shifts was 3,647 and the number of comparison shifts was 29,052. The c‐statistic for the MEWS(re) in the derivation dataset was 0.709 (0.6940.725); in the validation dataset, it was 0.698 (0.6920.714). The corresponding values for the EMR‐based models were 0.856 (0.8350.877) and 0.780 (0.7560.804). We also tested models in which, instead of dropping shifts with missing vital signs, we imputed missing vital signs to their normal value. The c‐statistic for the EMR‐based model with imputed vital sign values was 0.842 (0.8230.861) in the derivation dataset and 0.773 (0.7520.794) in the validation dataset. Lastly, we applied model coefficients to a dataset consisting of 4,290 randomly selected comparison shifts plus the 429 shifts excluded because of the 4‐hour length‐of‐stay criterion. The c‐statistic for this analysis was 0.756 (0.7030.809).
As a general rule, the EMR‐based models were more than twice as efficient as the MEWS(re). For example, a MEWS(re) threshold of 6 as the trigger for an alarm would identify 15% of all transfers to the ICU, with 34.4 false alarms for each transfer; in contrast, using the EMR‐based approach to identify 15% of all transfers, there were 14.5 false alarms for each transfer. Applied to the entire KPMCP Northern California Region, using the MEWS(re), a total of 52 patients per day would need to be evaluated, but only 22 per day using the EMR‐based approach. If one employed a MEWS(re) threshold of 4, this would lead to identification of 44% of all transfers, with a ratio of 69 false alarms for each transfer; using the EMR, the ratio would be 34 to 1. Across the entire KPMCP, a total of 276 patients per day (or about 19.5 a day per hospital) would need to be evaluated using the MEWS(re), but only 136 (or about 9.5 per hospital per day) using the EMR.
DISCUSSION
Using data from a large hospital cohort, we have developed a predictive model suitable for use in non‐ICU populations cared for in integrated healthcare settings with fully automated EMRs. The overall performance of our model, which incorporates acute physiology, diagnosis, and longitudinal data, is superior to the predictive ability of a model that can be assigned manually. This is not surprising, given that scoring systems such as the MEWS make an explicit tradeoff losing information found in multiple variables in exchange for ease of manual assignment. Currently, the model described in this report is being implemented in a simulated environment, a final safety test prior to piloting real‐time provision of probability estimates to clinicians and nurses. Though not yet ready for real‐time use, it is reasonable for our model to be tested using the KPHC shadow server, since evaluation in a simulated environment constitutes a critical evaluation step prior to deployment for clinical use. We also anticipate further refinement and revalidation to occur as more inpatient data become available in the KPMCP and elsewhere.
A number of limitations to our approach must be emphasized. In developing our models, we determined that, while modeling by clinical condition was important, the study outcome was rare for some primary conditions. In these diagnostic groups, which accounted for 12.5% of the event shifts and 10.6% of the comparison shifts, the c‐statistic in the validation dataset was <0.70. Since all 22 KPMCP hospitals are now online and will generate an additional 150,000 adult hospitalizations per year, we expect to be able to correct this problem prior to deployment of these models for clinical use. Having additional data will permit us to improve model discrimination and thus decrease the evaluation‐to‐detection ratio. In future iterations of these models, more experimentation with grouping of International Classification of Diseases (ICD) codes may be required. The problem of grouping ICD codes is not an easy one to resolve, in that diagnoses in the grouping must share common pathophysiology while having a grouping with a sufficient number of adverse events for stable statistical models.
Ideally, it would have been desirable to employ a more objective measure of deterioration, since the decision to transfer a patient to the ICU is discretionary. However, we have found that key data points needed to define such a measure (eg, vital signs) are not consistently charted when a patient deterioratesthis is not surprising outside the research setting, given that nurses and physicians involved in a transfer may be focusing on caring for the patient rather than immediately charting. Given the complexities of end‐of‐life‐care decision‐making, we could not employ death as the outcome of interest. A related issue is that our model does not differentiate between reasons for needing transfer to the ICU, an issue recently discussed by Bapoje et al.18
Our model does not address an important issue raised by Bapoje et al18 and Litvak, Pronovost, and others,19, 20 namely, whether a patient should have been admitted to a non‐ICU setting in the first place. Our team is currently developing a model for doing exactly this (providing decision support for triage in the emergency department), but discussion of this methodology is outside the scope of this article.
Because of resource and data limitations, our model also does not include newborns, children, women admitted for childbirth, or patients transferred from non‐KPMCP hospitals. However, the approach described here could serve as a starting point for developing models for these other populations.
The generalizability of our model must also be considered. The Northern California KPMCP is unusual in having large electronic databases that include physiologic as well as longitudinal patient data. Many hospitals cannot take advantage of all the methods described here. However, the methods we employed could be modified for use by hospital systems in countries such as Great Britain and Canada, and entities such as the Veterans Administration Hospital System in the United States. The KPMCP population, an insured population with few barriers to access, is healthier than the general population, and some population subsets are underrepresented in our cohort. Practice patterns may also vary. Nonetheless, the model described here could serve as a good starting point for future collaborative studies, and it would be possible to develop models suitable for use by stand‐alone hospitals (eg, recalibrating so that one used a Charlson comorbidity21 score based on present on‐admission codes rather than the COPS).
The need for early detection of patient deterioration has played a major role in the development of rapid response teams, as well as scores such as the MEWS. In particular, entities such as the Institute for Healthcare Improvement have advocated the use of early warning systems.22 However, having a statistically robust model to support an early warning system is only part of the solution, and a number of new challenges must then be addressed. The first is actual electronic deployment. Existing inpatient EMRs were not designed with complex calculations in mind, and we anticipate that some degradation in performance will occur when we test our models using real‐time data capture. As Bapoje et al point out, simply having an alert may be insufficient, since not all transfers are preventable.18 Early warning systems also raise ethical issues (for example, what should be done if an alert leads a clinician to confront the fact that an end‐of‐life‐care discussion needs to occur?). From a research perspective, if one were to formally test the benefits of such models, it would be critical to define outcome measures other than death (which is strongly affected by end‐of‐life‐care decisions) or ICU transfer (which is often desirable).
In conclusion, we have developed an approach for predicting impending physiologic deterioration of hospitalized adults outside the ICU. Our approach illustrates how organizations can take maximal advantage of EMRs in a manner that exceeds meaningful use specifications.23, 24 Our study highlights the possibility of using fully automated EMR data for building and applying sophisticated statistical models in settings other than the highly monitored ICU without the need for additional equipment. It also expands the universe of severity scoring to one in which probability estimates are provided in real time and throughout an entire hospitalization. Model performance will undoubtedly improve over time, as more patient data become available. Although our approach has important limitations, it is suitable for testing using real‐time data in a simulated environment. Such testing would permit identification of unanticipated problems and quantification of the degradation of model performance due to real life factors, such as delays in vital signs charting or EMR system brownouts. It could also serve as the springboard for future collaborative studies, with a broader population base, in which the EMR becomes a tool for care, not just documentation.
Acknowledgements
We thank Ms Marla Gardner and Mr John Greene for their work in the development phase of this project. We are grateful to Brian Hoberman, Andrew Hwang, and Marc Flagg from the RIMS group; to Colin Stobbs, Sriram Thiruvenkatachari, and Sundeep Sood from KP IT, Inc; and to Dennis Andaya, Linda Gliner, and Cyndi Vasallo for their assistance with data‐quality audits. We are also grateful to Dr Philip Madvig, Dr Paul Feigenbaum, Dr Alan Whippy, Mr Gregory Adams, Ms Barbara Crawford, and Dr Marybeth Sharpe for their administrative support and encouragement; and to Dr Alan S. Go, Acting Director of the Kaiser Permanente Division of Research, for reviewing the manuscript.
Patients in general medicalsurgical wards who experience unplanned transfer to the intensive care unit (ICU) have increased mortality and morbidity.13 Using an externally validated methodology permitting assessment of illness severity and mortality risk among all hospitalized patients,4, 5 we recently documented observed‐to‐expected mortality ratios >3.0 and excess length of stay of 10 days among patients who experienced such transfers.6
It is possible to predict adverse outcomes among monitored patients (eg, patients in the ICU or undergoing continuous electronic monitoring).7, 8 However, prediction of unplanned transfers among medicalsurgical ward patients presents challenges. Data collection (vital signs and laboratory tests) is relatively infrequent. The event rate (3% of hospital admissions) is low, and the rate in narrow time periods (eg, 12 hours) is extremely low: a hospital with 4000 admissions per year might experience 1 unplanned transfer to the ICU every 3 days. Not surprisingly, performance of models suitable for predicting ward patients' need for intensive care within narrow time frames have been disappointing.9 The Modified Early Warning Score (MEWS), has a c‐statistic, or area under the receiver operator characteristic of 0.67,1012 and our own model incorporating 14 laboratory tests, but no vital signs, has excellent performance with respect to predicting inpatient mortality, but poor performance with respect to unplanned transfer.6
In this report, we describe the development and validation of a complex predictive model suitable for use with ward patients. Our objective for this work was to develop a predictive model based on clinical and physiologic data available in real time from a comprehensive electronic medical record (EMR), not a clinically intuitive, manually assigned tool. The outcome of interest was unplanned transfer from the ward to the ICU, or death on the ward in a patient who was full code. This model has been developed as part of a regional effort to decrease preventable mortality in the Northern California Kaiser Permanente Medical Care Program (KPMCP), an integrated healthcare delivery system with 22 hospitals.
MATERIALS AND METHODS
For additional details, see the Supporting Information, Appendices 112, in the online version of this article.
This project was approved by the KPMCP Institutional Board for the Protection of Human Subjects.
The Northern California KPMCP serves a total population of approximately 3.3 million members. All Northern California KPMCP hospitals and clinics employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere. Databases maintained by the KPMCP capture admission and discharge times, admission and discharge diagnoses and procedures (assigned by professional coders), bed histories permitting quantification of intra‐hospital transfers, inter‐hospital transfers, as well as the results of all inpatient and outpatient laboratory tests. In July 2006, the KPMCP began deployment of the EMR developed by Epic Systems Corporation (
Our setting consisted of 14 hospitals in which the KPHC inpatient EMR had been running for at least 3 months (the KPMCP Antioch, Fremont, Hayward, Manteca, Modesto, Roseville, Sacramento, Santa Clara, San Francisco, Santa Rosa, South Sacramento, South San Francisco, Santa Teresa, and Walnut Creek hospitals). We have described the general characteristics of KPMCP hospitals elsewhere.4, 6 Our initial study population consisted of all patients admitted to these hospitals who met the following criteria: hospitalization began from November 1, 2006 through December 31, 2009; initial hospitalization occurred at a Northern California KPMCP hospital (ie, for inter‐hospital transfers, the first hospital stay occurred within the KPMCP); age 18 years; hospitalization was not for childbirth; and KPHC had been operational at the hospital for at least 3 months.
Analytic Approach
The primary outcome for this study was transfer to the ICU after admission to the hospital among patients residing either in a general medicalsurgical ward (ward) or transitional care unit (TCU), or death in the ward or TCU in a patient who was full code at the time of death (ie, had the patient survived, s/he would have been transferred to the ICU). The unit of analysis for this study was a 12‐hour patient shift, which could begin with a 7 AM T0 (henceforth, day shift) or a 7 PM T0 (night shift); in other words, we aimed to predict the occurrence of an event within 12 hours of T0 using only data available prior to T0. A shift in which a patient experienced the primary study outcome is an event shift, while one in which a patient did not experience the primary outcome is a comparison shift. Using this approach, an individual patient record could consist of both event and comparison shifts, since some patients might have multiple unplanned transfers and some patients might have none. Our basic analytic approach consisted of creating a cohort of event and comparison shifts (10 comparison shifts were randomly selected for each event shift), splitting the cohort into a derivation dataset (50%) and validation dataset (50%), developing a model using the derivation dataset, then applying the coefficients of the derivation dataset to the validation dataset. Because some event shifts were excluded due to the minimum 4‐hour length‐of‐stay requirement, we also applied model coefficients to these excluded shifts and a set of randomly selected comparison shifts.
Since the purpose of these analyses was to develop models with maximal signal extraction from sparsely collected predictors, we did not block a time period after the T0 to allow for a reaction time to the alarm. Thus, since some events could occur immediately after the T0 (as can be seen in the Supporting Information, Appendices, in the online version of this article), our models would need to be run at intervals that are more frequent than 2 times a day.
Independent Variables
In addition to patients' age and sex, we tested the following candidate independent variables. Some of these variables are part of the KPMCP risk adjustment model4, 5 and were available electronically for all patients in the cohort. We grouped admission diagnoses into 44 broad diagnostic categories (primary conditions), and admission types into 4 groups (emergency medical, emergency surgical, elective medical, and elective surgical). We quantified patients' degree of physiologic derangement in the 72 hours preceding hospitalization with a Laboratory‐based Acute Physiology Score (LAPS) using 14 laboratory test results prior to hospitalization; we also tested individual laboratory test results obtained after admission to the hospital. We quantified patients' comorbid illness burden using a COmorbidity Point Score (COPS) based on patients' preexisting diagnoses over the 12‐month period preceding hospitalization.4 We extracted temperature, heart rate, respiratory rate, systolic blood pressure, diastolic blood pressure, oxygen saturation, and neurological status from the EMR. We also tested the following variables based on specific information extracted from the EMR: shock index (heart rate divided by systolic blood pressure)13; care directive status (patients were placed into 4 groups: full code, partial code, do not resuscitate [DNR], and no care directive in place); and a proxy for measured lactate (PML; anion gap/serum bicarbonate 100).1416 For comparison purposes, we also created a retrospective electronically assigned MEWS, which we refer to as the MEWS(re), and we assigned this score to patient records electronically using data from KP HealthConnect.
Statistical Methods
Analyses were performed in SAS 9.1, Stata 10, and R 2.12. Final validation was performed using SAS (SAS Institute Inc., Carey, North Carolina). Since we did not limit ourselves to traditional severity‐scoring approaches (eg, selecting the worst heart rate in a given time interval), but also included trend terms (eg, change in heart rate over the 24 hours preceding T0), the number of potential variables to test was very large. Detailed description of the statistical strategies employed for variable selection is provided in the Supporting Information, Appendices, in the online version of this article. Once variables were selected, our basic approach was to test a series of diagnosis‐specific logistic regression submodels using a variety of predictors that included vital signs, vital signs trends (eg, most recent heart rate minus earliest heart rate, heart rate over preceding 24 hours), and other above‐mentioned variables.
We assessed the ability of a submodel to correctly distinguish patients who died, from survivors, using the c‐statistic, as well as other metrics recommended by Cook.17 At the end of the modeling process, we pooled the results across all submodels. For vital signs, where the rate of missing data was <3%, we tested submodels in which we dropped shifts with missing data, as well as submodels in which we imputed missing vital signs to a normal value. For laboratory data, where the rate of missing data for a given shift was much greater, we employed a probabilistic imputation method that included consideration of when a laboratory test result became available.
RESULTS
During the study period, a total of 102,488 patients experienced 145,335 hospitalizations at the study hospitals. We removed 66 patients with 138 hospitalizations for data quality reasons, leaving us with our initial study sample of 102,422 patients whose characteristics are summarized in Table 1. Table 1, in which the unit of analysis is an individual patient, shows that patients who experienced the primary outcome were similar to those patients described in our previous report, in terms of their characteristics on admission as well as in experiencing excess morbidity and mortality.6
Never Admitted to ICU | Direct Admit to ICU From ED | Unplanned Transfer to ICU* | Other ICU Admission | |
---|---|---|---|---|
| ||||
N | 89,269 | 5963 | 2880 | 4310 |
Age (mean SD) | 61.26 18.62 | 62.25 18.13 | 66.12 16.20 | 64.45 15.91 |
Male (n, %) | 37,228 (41.70%) | 3091 (51.84%) | 1416 (49.17%) | 2378 (55.17%) |
LAPS (mean SD) | 13.02 15.79 | 32.72 24.85 | 24.83 21.53 | 11.79 18.16 |
COPS(mean SD) | 67.25 51.42 | 73.88 57.42 | 86.33 59.33 | 78.44 52.49 |
% Predicted mortality risk (mean SD) | 1.93% 3.98% | 7.69% 12.59% | 5.23% 7.70% | 3.66% 6.81% |
Survived first hospitalization to discharge∥ | 88,479 (99.12%) | 5336 (89.49%) | 2316 (80.42%) | 4063 (94.27%) |
Care order on admission | ||||
Full code | 78,877 (88.36%) | 5198 (87.17%) | 2598 (90.21%) | 4097 (95.06%) |
Partial code | 664 (0.74%) | 156 (2.62%) | 50 (1.74%) | 27 (0.63%) |
Comfort care | 21 (0.02%) | 2 (0.03%) | 0 (0%) | 0 (0%) |
DNR | 8227 (9.22%) | 539 (9.04%) | 219 (7.60%) | 161 (3.74%) |
Comfort care and DNR | 229 (0.26%) | 9 (0.15%) | 2 (0.07%) | 2 (0.05%) |
No order | 1251 (1.40%) | 59 (0.99%) | 11 (0.38%) | 23 (0.53%) |
Admission diagnosis (n, %) | ||||
Pneumonia | 2385 (2.67%) | 258 (4.33%) | 242 (8.40%) | 68 (1.58%) |
Sepsis | 5822 (6.52%) | 503 (8.44%) | 279 (9.69%) | 169 (3.92%) |
GI bleeding | 9938 (11.13%) | 616 (10.33%) | 333 (11.56%) | 290 (6.73%) |
Cancer | 2845 (3.19%) | 14 (0.23%) | 95 (3.30%) | 492 (11.42%) |
Total hospital length of stay (days SD) | 3.08 3.29 | 5.37 7.50 | 12.16 13.12 | 8.06 9.53 |
Figure 1shows how we developed the analysis cohort, by removing patients with a comfort‐care‐only order placed within 4 hours after admission (369 patients/744 hospitalizations) and patients who were never admitted to the ward or TCU (7,220/10,574). This left a cohort consisting of 94,833 patients who experienced 133,879 hospitalizations spanning a total of 1,079,062 shifts. We then removed shifts where: 1) a patient was not on the ward at the start of a shift, or was on the ward for <4 hours of a shift; 2) the patient had a comfort‐care order in place at the start of the shift; and 3) the patient died and was ineligible to be a case (the patient had a DNR order in place or died in the ICU). The final cohort eligible for sampling consisted of 846,907 shifts, which involved a total of 92,797 patients and 130,627 hospitalizations. There were a total of 4,036 event shifts, which included 3,224 where a patient was transferred from the ward to the ICU, 717 from the TCU to the ICU, and 95 where a patient died on the ward or TCU without a DNR order in place. We then randomly selected 39,782 comparison shifts. Thus, our final cohort for analysis included 4,036 event shifts (1,979 derivation/2,057 validation and 39,782 comparison shifts (19,509/20,273). As a secondary validation, we also applied model coefficients to the 429 event shifts excluded due to the <4‐hour length‐of‐stay requirement.
Table 2 compares event shifts with comparison shifts. In the 24 hours preceding ICU transfer, patients who were subsequently transferred had statistically significant, but not necessarily clinically significant, differences in terms of these variables. However, missing laboratory data were more common, ranging from 18% to 31% of all shifts (we did not incorporate laboratory tests where 35% of the shifts had missing data for that test).
Predictor | Event Shifts | Comparison Shifts | P |
---|---|---|---|
| |||
Number | 4036 | 39,782 | |
Age (mean SD) | 67.19 15.25 | 65.41 17.40 | <0.001 |
Male (n, %) | 2007 (49.73%) | 17,709 (44.52%) | <0.001 |
Day shift | 1364 (33.80%) | 17,714 (44.53%) | <0.001 |
LAPS* | 27.89 22.10 | 20.49 20.16 | <0.001 |
COPS | 116.33 72.31 | 100.81 68.44 | <0.001 |
Full code (n, %) | 3496 (86.2%) | 32,156 (80.8%) | <0.001 |
ICU shift during hospitalization | 3964 (98.22%) | 7197 (18.09%) | <0.001 |
Unplanned transfer to ICU during hospitalization∥ | 353 (8.8%) | 1466 (3.7%) | <0.001 |
Temperature (mean SD) | 98.15 (1.13) | 98.10 (0.85) | 0.009 |
Heart rate (mean SD) | 90.30 (20.48) | 79.86 (5.27) | <0.001 |
Respiratory rate (mean SD) | 20.36 (3.70) | 18.87 (1.79) | <0.001 |
Systolic blood pressure (mean SD) | 123.65 (23.26) | 126.21 (19.88) | <0.001 |
Diastolic blood pressure (mean SD) | 68.38 (14.49) | 69.46 (11.95) | <0.001 |
Oxygen saturation (mean SD) | 95.72% (3.00) | 96.47 % (2.26) | <0.001 |
MEWS(re) (mean SD) | 3.64 (2.02) | 2.34 (1.61) | <0.001 |
% <5 | 74.86% | 92.79% | |
% 5 | 25.14% | 7.21% | <0.001 |
Proxy for measured lactate# (mean SD) | 36.85 (28.24) | 28.73 (16.74) | <0.001 |
% Missing in 24 hr before start of shift** | 17.91% | 28.78% | <0.001 |
Blood urea nitrogen (mean SD) | 32.03 (25.39) | 22.72 (18.9) | <0.001 |
% Missing in 24 hr before start of shift | 19.67% | 20.90% | <0.001 |
White blood cell count 1000 (mean SD) | 12.33 (11.42) | 9.83 (6.58) | <0.001 |
% Missing in 24 hr before start of shift | 21.43% | 30.98% | <0.001 |
Hematocrit (mean SD) | 33.08 (6.28) | 33.07 (5.25) | 0.978 |
% Missing in 24 hr before start of shift | 19.87% | 29.55% | <0.001 |
After conducting multiple analyses using the derivation dataset, we developed 24 submodels, a compromise between our finding that primary‐condition‐specific models showed better performance and the fact that we had very few events among patients with certain primary conditions (eg, pericarditis/valvular heart disease), which forced us to create composite categories (eg, a category pooling patients with pericarditis, atherosclerosis, and peripheral vascular disease). Table 3 lists variables included in our final submodels.
Variable | Description |
---|---|
| |
Directive status | Full code or not full code |
LAPS* | Admission physiologic severity of illness score (continuous variable ranging from 0 to 256). Standardized and included as LAPS and LAPS squared |
COPS | Comorbidity burden score (continuous variable ranging from 0 to 701). Standardized and included as COPS and COPS squared. |
COPS status | Indicator for absent comorbidity data |
LOS at T0 | Length of stay in the hospital (total time in hours) at the T0; standardized. |
T0 time of day | 7 AM or 7 PM |
Temperature | Worst (highest) temperature in 24 hr preceding T0; variability in temperature in 24 hr preceding T0. |
Heart rate | Most recent heart rate in 24 hr preceding T0; variability in heart rate in 24 hr preceding T0. |
Respiratory rate | Most recent respiratory rate in 24 hr preceding T0; worst (highest) respiratory rate in 24 hr preceding T0; variability in respiratory rate in 24 hr preceding T0. |
Diastolic blood pressure | Most recent diastolic blood pressure in 24 hr preceding T0 transformed by subtracting 70 from the actual value and squaring the result. Any value above 2000 is subsequently then set to 2000, yielding a continuous variable ranging from 0 to 2000. |
Systolic pressure | Variability in systolic blood pressure in 24 hr preceding T0. |
Pulse oximetry | Worst (lowest) oxygen saturation in 24 hr preceding T0; variability in oxygen saturation in 24 hr preceding T0. |
Neurological status | Most recent neurological status check in 24 hr preceding T0. |
Laboratory tests | Blood urea nitrogen |
Proxy for measured lactate = (anion gap serum bicarbonate) 100 | |
Hematocrit | |
Total white blood cell count |
Table 4 summarizes key results in the validation dataset. Across all diagnoses, the MEWS(re) had c‐statistic of 0.709 (95% confidence interval, 0.6970.721) in the derivation dataset and 0.698 (0.6860.710) in the validation dataset. In the validation dataset, the MEWS(re) performed best among patients with a set of gastrointestinal diagnoses (c = 0.792; 0.7260.857) and worst among patients with congestive heart failure (0.541; 0.5000.620). In contrast, across all primary conditions, the EMR‐based models had a c‐statistic of 0.845 (0.8260.863) in the derivation dataset and 0.775 (0.7530.797) in the validation dataset. In the validation dataset, the EMR‐based models also performed best among patients with a set of gastrointestinal diagnoses (0.841; 0.7830.897) and worst among patients with congestive heart failure (0.683; 0.6100.755). A negative correlation (R = 0.63) was evident between the number of event shifts in a submodel and the drop in the c‐statistic seen in the validation dataset.
No. of Shifts in Validation Dataset | c‐Statistic | |||
---|---|---|---|---|
Diagnoses Group* | Event | Comparison | MEWS(re) | EMR Model |
| ||||
Acute myocardial infarction | 36 | 169 | 0.541 | 0.572 |
Diseases of pulmonary circulation and cardiac dysrhythmias | 40 | 329 | 0.565 | 0.645 |
Seizure disorders | 45 | 497 | 0.594 | 0.647 |
Rule out myocardial infarction | 77 | 727 | 0.602 | 0.648 |
Pneumonia | 163 | 847 | 0.741 | 0.801 |
GI diagnoses, set A | 58 | 942 | 0.755 | 0.803 |
GI diagnoses, set B∥ | 256 | 2,610 | 0.772 | 0.806 |
GI diagnoses, set C | 46 | 520 | 0.792 | 0.841 |
All diagnosis | 2,032 | 20,106 | 0.698 | 0.775 |
We also compared model performance when our datasets were restricted to 1 randomly selected observation per patient; in these analyses, the total number of event shifts was 3,647 and the number of comparison shifts was 29,052. The c‐statistic for the MEWS(re) in the derivation dataset was 0.709 (0.6940.725); in the validation dataset, it was 0.698 (0.6920.714). The corresponding values for the EMR‐based models were 0.856 (0.8350.877) and 0.780 (0.7560.804). We also tested models in which, instead of dropping shifts with missing vital signs, we imputed missing vital signs to their normal value. The c‐statistic for the EMR‐based model with imputed vital sign values was 0.842 (0.8230.861) in the derivation dataset and 0.773 (0.7520.794) in the validation dataset. Lastly, we applied model coefficients to a dataset consisting of 4,290 randomly selected comparison shifts plus the 429 shifts excluded because of the 4‐hour length‐of‐stay criterion. The c‐statistic for this analysis was 0.756 (0.7030.809).
As a general rule, the EMR‐based models were more than twice as efficient as the MEWS(re). For example, a MEWS(re) threshold of 6 as the trigger for an alarm would identify 15% of all transfers to the ICU, with 34.4 false alarms for each transfer; in contrast, using the EMR‐based approach to identify 15% of all transfers, there were 14.5 false alarms for each transfer. Applied to the entire KPMCP Northern California Region, using the MEWS(re), a total of 52 patients per day would need to be evaluated, but only 22 per day using the EMR‐based approach. If one employed a MEWS(re) threshold of 4, this would lead to identification of 44% of all transfers, with a ratio of 69 false alarms for each transfer; using the EMR, the ratio would be 34 to 1. Across the entire KPMCP, a total of 276 patients per day (or about 19.5 a day per hospital) would need to be evaluated using the MEWS(re), but only 136 (or about 9.5 per hospital per day) using the EMR.
DISCUSSION
Using data from a large hospital cohort, we have developed a predictive model suitable for use in non‐ICU populations cared for in integrated healthcare settings with fully automated EMRs. The overall performance of our model, which incorporates acute physiology, diagnosis, and longitudinal data, is superior to the predictive ability of a model that can be assigned manually. This is not surprising, given that scoring systems such as the MEWS make an explicit tradeoff losing information found in multiple variables in exchange for ease of manual assignment. Currently, the model described in this report is being implemented in a simulated environment, a final safety test prior to piloting real‐time provision of probability estimates to clinicians and nurses. Though not yet ready for real‐time use, it is reasonable for our model to be tested using the KPHC shadow server, since evaluation in a simulated environment constitutes a critical evaluation step prior to deployment for clinical use. We also anticipate further refinement and revalidation to occur as more inpatient data become available in the KPMCP and elsewhere.
A number of limitations to our approach must be emphasized. In developing our models, we determined that, while modeling by clinical condition was important, the study outcome was rare for some primary conditions. In these diagnostic groups, which accounted for 12.5% of the event shifts and 10.6% of the comparison shifts, the c‐statistic in the validation dataset was <0.70. Since all 22 KPMCP hospitals are now online and will generate an additional 150,000 adult hospitalizations per year, we expect to be able to correct this problem prior to deployment of these models for clinical use. Having additional data will permit us to improve model discrimination and thus decrease the evaluation‐to‐detection ratio. In future iterations of these models, more experimentation with grouping of International Classification of Diseases (ICD) codes may be required. The problem of grouping ICD codes is not an easy one to resolve, in that diagnoses in the grouping must share common pathophysiology while having a grouping with a sufficient number of adverse events for stable statistical models.
Ideally, it would have been desirable to employ a more objective measure of deterioration, since the decision to transfer a patient to the ICU is discretionary. However, we have found that key data points needed to define such a measure (eg, vital signs) are not consistently charted when a patient deterioratesthis is not surprising outside the research setting, given that nurses and physicians involved in a transfer may be focusing on caring for the patient rather than immediately charting. Given the complexities of end‐of‐life‐care decision‐making, we could not employ death as the outcome of interest. A related issue is that our model does not differentiate between reasons for needing transfer to the ICU, an issue recently discussed by Bapoje et al.18
Our model does not address an important issue raised by Bapoje et al18 and Litvak, Pronovost, and others,19, 20 namely, whether a patient should have been admitted to a non‐ICU setting in the first place. Our team is currently developing a model for doing exactly this (providing decision support for triage in the emergency department), but discussion of this methodology is outside the scope of this article.
Because of resource and data limitations, our model also does not include newborns, children, women admitted for childbirth, or patients transferred from non‐KPMCP hospitals. However, the approach described here could serve as a starting point for developing models for these other populations.
The generalizability of our model must also be considered. The Northern California KPMCP is unusual in having large electronic databases that include physiologic as well as longitudinal patient data. Many hospitals cannot take advantage of all the methods described here. However, the methods we employed could be modified for use by hospital systems in countries such as Great Britain and Canada, and entities such as the Veterans Administration Hospital System in the United States. The KPMCP population, an insured population with few barriers to access, is healthier than the general population, and some population subsets are underrepresented in our cohort. Practice patterns may also vary. Nonetheless, the model described here could serve as a good starting point for future collaborative studies, and it would be possible to develop models suitable for use by stand‐alone hospitals (eg, recalibrating so that one used a Charlson comorbidity21 score based on present on‐admission codes rather than the COPS).
The need for early detection of patient deterioration has played a major role in the development of rapid response teams, as well as scores such as the MEWS. In particular, entities such as the Institute for Healthcare Improvement have advocated the use of early warning systems.22 However, having a statistically robust model to support an early warning system is only part of the solution, and a number of new challenges must then be addressed. The first is actual electronic deployment. Existing inpatient EMRs were not designed with complex calculations in mind, and we anticipate that some degradation in performance will occur when we test our models using real‐time data capture. As Bapoje et al point out, simply having an alert may be insufficient, since not all transfers are preventable.18 Early warning systems also raise ethical issues (for example, what should be done if an alert leads a clinician to confront the fact that an end‐of‐life‐care discussion needs to occur?). From a research perspective, if one were to formally test the benefits of such models, it would be critical to define outcome measures other than death (which is strongly affected by end‐of‐life‐care decisions) or ICU transfer (which is often desirable).
In conclusion, we have developed an approach for predicting impending physiologic deterioration of hospitalized adults outside the ICU. Our approach illustrates how organizations can take maximal advantage of EMRs in a manner that exceeds meaningful use specifications.23, 24 Our study highlights the possibility of using fully automated EMR data for building and applying sophisticated statistical models in settings other than the highly monitored ICU without the need for additional equipment. It also expands the universe of severity scoring to one in which probability estimates are provided in real time and throughout an entire hospitalization. Model performance will undoubtedly improve over time, as more patient data become available. Although our approach has important limitations, it is suitable for testing using real‐time data in a simulated environment. Such testing would permit identification of unanticipated problems and quantification of the degradation of model performance due to real life factors, such as delays in vital signs charting or EMR system brownouts. It could also serve as the springboard for future collaborative studies, with a broader population base, in which the EMR becomes a tool for care, not just documentation.
Acknowledgements
We thank Ms Marla Gardner and Mr John Greene for their work in the development phase of this project. We are grateful to Brian Hoberman, Andrew Hwang, and Marc Flagg from the RIMS group; to Colin Stobbs, Sriram Thiruvenkatachari, and Sundeep Sood from KP IT, Inc; and to Dennis Andaya, Linda Gliner, and Cyndi Vasallo for their assistance with data‐quality audits. We are also grateful to Dr Philip Madvig, Dr Paul Feigenbaum, Dr Alan Whippy, Mr Gregory Adams, Ms Barbara Crawford, and Dr Marybeth Sharpe for their administrative support and encouragement; and to Dr Alan S. Go, Acting Director of the Kaiser Permanente Division of Research, for reviewing the manuscript.
- Day of the week of intensive care admission and patient outcomes: a multisite regional evaluation.Med Care.2002;40(6):530–539. , , , .
- The hospital mortality of patients admitted to the ICU on weekends.Chest.2004;126(4):1292–1298. , , , et al.
- Mortality among patients admitted to intensive care units during weekday day shifts compared with “off” hours.Crit Care Med.2007;35(1):3–11. , , , et al.
- Risk adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases.Med Care.2008;46(3):232–239. , , , , , .
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population.J Clin Epidemiol.2010;63(7):798–803. , , , .
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS).J Hosp Med.2011;6(2):74–80. , , , , , .
- Multicentric study of monitoring alarms in the adult intensive care unit (ICU): a descriptive analysis.Intensive Care Med.1999;25(12):1360–1366. , , , , , .
- Integration of early physiological responses predicts later illness severity in preterm infants.Sci Transl Med.2010;2(48):48ra65. , , , , .
- Reproducibility of physiological track‐and‐trigger warning systems for identifying at‐risk patients on the ward.Intensive Care Med.2007;33(4):619–624. , , .
- Validation of a Modified Early Warning Score in medical admissions.Q J Med.2001;94:521–526. , , , .
- Effect of introducing the Modified Early Warning score on clinical outcomes, cardio‐pulmonary arrests and intensive care utilisation in acute medical admissions.Anaesthesia.2003;58(8):797–802. , , , , .
- MERIT Study Investigators.Introduction of the medical emergency team (MET) system: a cluster‐randomized controlled trial.Lancet.2005;365(9477):2091–2097.
- Unplanned transfers to the intensive care unit: the role of the shock index.J Hosp Med.2010;5(8):460–465. , , , , , .
- The delta (delta) gap: an approach to mixed acid‐base disorders.Ann Emerg Med.1990;19(11):1310–1313. .
- Acid‐base disorders: classification and management strategies.Am Fam Physician.1995;52(2):584–590. .
- Unmeasured anions in critically ill patients: can they predict mortality?Crit Care Med.2003;31(8):2131–2136. , , , .
- Use and misuse of the receiver operating characteristic curve in risk prediction.Circulation.2007;115(7):928–935. .
- Unplanned transfers to a medical intensive care unit: causes and relationship to preventable errors in care.J Hosp Med.2011;6(2):68–72. , , , .
- Rethinking rapid response teams.JAMA.2010;304(12):1375–1376. , .
- Rapid response teams—walk, don't run.JAMA.2006;296(13):1645–1647. , , .
- A new method of classifying prognostic comorbidity in longitudinal populations: development and validation.J Chronic Dis.1987;40:373–383. , , , .
- Institute for Healthcare Improvement.Early Warning Systems:The Next Level of Rapid Response.2011. http://www.ihi.org/IHI/Programs/AudioAndWebPrograms/ExpeditionEarlyWarningSystemsTheNextLevelofRapidResponse.htm?player=wmp. Accessed 4/6/11.
- Assessing readiness for meeting meaningful use: identifying electronic health record functionality and measuring levels of adoption.AMIA Annu Symp Proc.2010;2010:66–70. .
- Medicare and Medicaid Programs;Electronic Health Record Incentive Program. Final Rule.Fed Reg.2010;75(144):44313–44588.
- Day of the week of intensive care admission and patient outcomes: a multisite regional evaluation.Med Care.2002;40(6):530–539. , , , .
- The hospital mortality of patients admitted to the ICU on weekends.Chest.2004;126(4):1292–1298. , , , et al.
- Mortality among patients admitted to intensive care units during weekday day shifts compared with “off” hours.Crit Care Med.2007;35(1):3–11. , , , et al.
- Risk adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases.Med Care.2008;46(3):232–239. , , , , , .
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population.J Clin Epidemiol.2010;63(7):798–803. , , , .
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS).J Hosp Med.2011;6(2):74–80. , , , , , .
- Multicentric study of monitoring alarms in the adult intensive care unit (ICU): a descriptive analysis.Intensive Care Med.1999;25(12):1360–1366. , , , , , .
- Integration of early physiological responses predicts later illness severity in preterm infants.Sci Transl Med.2010;2(48):48ra65. , , , , .
- Reproducibility of physiological track‐and‐trigger warning systems for identifying at‐risk patients on the ward.Intensive Care Med.2007;33(4):619–624. , , .
- Validation of a Modified Early Warning Score in medical admissions.Q J Med.2001;94:521–526. , , , .
- Effect of introducing the Modified Early Warning score on clinical outcomes, cardio‐pulmonary arrests and intensive care utilisation in acute medical admissions.Anaesthesia.2003;58(8):797–802. , , , , .
- MERIT Study Investigators.Introduction of the medical emergency team (MET) system: a cluster‐randomized controlled trial.Lancet.2005;365(9477):2091–2097.
- Unplanned transfers to the intensive care unit: the role of the shock index.J Hosp Med.2010;5(8):460–465. , , , , , .
- The delta (delta) gap: an approach to mixed acid‐base disorders.Ann Emerg Med.1990;19(11):1310–1313. .
- Acid‐base disorders: classification and management strategies.Am Fam Physician.1995;52(2):584–590. .
- Unmeasured anions in critically ill patients: can they predict mortality?Crit Care Med.2003;31(8):2131–2136. , , , .
- Use and misuse of the receiver operating characteristic curve in risk prediction.Circulation.2007;115(7):928–935. .
- Unplanned transfers to a medical intensive care unit: causes and relationship to preventable errors in care.J Hosp Med.2011;6(2):68–72. , , , .
- Rethinking rapid response teams.JAMA.2010;304(12):1375–1376. , .
- Rapid response teams—walk, don't run.JAMA.2006;296(13):1645–1647. , , .
- A new method of classifying prognostic comorbidity in longitudinal populations: development and validation.J Chronic Dis.1987;40:373–383. , , , .
- Institute for Healthcare Improvement.Early Warning Systems:The Next Level of Rapid Response.2011. http://www.ihi.org/IHI/Programs/AudioAndWebPrograms/ExpeditionEarlyWarningSystemsTheNextLevelofRapidResponse.htm?player=wmp. Accessed 4/6/11.
- Assessing readiness for meeting meaningful use: identifying electronic health record functionality and measuring levels of adoption.AMIA Annu Symp Proc.2010;2010:66–70. .
- Medicare and Medicaid Programs;Electronic Health Record Incentive Program. Final Rule.Fed Reg.2010;75(144):44313–44588.
Copyright © 2012 Society of Hospital Medicine
Outcomes of Delayed ICU Transfer
Hospitalized patients who require transfer from medical wards to the intensive care unit (ICU) have high in‐hospital mortality, in some reports exceeding 55%.14 In a previous report in this journal, we found that while these unplanned ICU transfers occurred in only 4% of hospitalizations, they were present in nearly one‐quarter of fatal hospitalizations and were associated with substantial increases in resource utilization.4 For these reasons, interventions aimed at identifying and treating this high‐risk group have received considerable attention and have been proposed as measures of inpatient safety.2, 49
Notably, mortality among patients with unplanned ICU transfers exceeds mortality among patients admitted to the ICU directly from the emergency department (ED)a group traditionally considered to have the highest risk of death.13, 10 Previous single‐center studies suggest that increased mortality rates are present even among patients transferred within 24 hours of hospital admission, and reinforce the notion that earlier recognition of critical illness may result in improved outcomes.1113 However, these studies have been performed primarily in small cohorts of heterogeneous patients, and may obscure the independent effect of unplanned transfers on mortality and hamper efforts to use unplanned transfer rates as a metric of healthcare quality.1, 2, 4, 9
In this study, we evaluated early unplanned ICU transfers drawn from a cohort of 499,995 hospitalizations in an integrated healthcare delivery system. Using patient data, extracted from the automated electronic medical record, we matched unplanned transfer cases to patients directly admitted to the ICU and described the association between delayed ICU transfers and adverse outcomes.
METHODS
Setting and Participants
We performed a retrospective analysis of adult patient (age 18 years) hospitalizations at 21 Northern California Kaiser Permanente (KP) Medical Care Program hospitals between January 2007 and December 2009. This work expanded on our previous report of hospital stays from November 2006 to January 2008.4 The 21 study hospitals used the same electronic health information systems; databases captured admission, discharge, and bed history data. The use of these databases for research has been described in our previous study and other reports; hospital characteristics, unit staffing, and resource levels have also been detailed previously.4, 1417 This study was approved by the KP Institutional Review Board.
Identifying Unplanned Transfers
We evaluated patients with medical hospitalizationsdefined as those whose first hospital location was not in a surgical setting such as the operating room or post‐anesthesia recovery areawhose admission originated in the ED; patients admitted for surgery were removed because of significant differences in observed mortality (see Supporting Information Appendix Figure 1 and Appendix Table 1 in the online version of this article). Patients whose admission did not originate in the ED were excluded to eliminate confounding resulting from differences in preadmission care. We also excluded patients admitted for gynecological and pregnancy‐related care because of low hospital mortality.
Initial patient locations included the medical wards (wards); the transitional care unit (TCU); and the intensive care unit (ICU). Bed history data, based on time stamps and available for all patients, were used to track patient locations from the time of admission, defined as the first non‐ED hospital location, until discharge. Patient length of stay (LOS) was calculated at each location and for the entire hospitalization.
Transfers to the ICU after a patient's initial admission to the ward or TCU were termed unplanned (or delayed) ICU transfers; patients admitted from the ED to the ICU were termed direct ICU admit patients. Direct ICU admit patients were excluded from the unplanned transfer group even if they required a readmission to the ICU later in their hospital course. We focused on patients with unplanned ICU transfers early after hospitalization to identify those in whom prompt recognition and intervention could be effective; thus, our primary analyses were on patients with transfers within 24 hours of admission. In secondary analysis, we also evaluated patients with unplanned ICU transfers occurring within 48 hours after hospital admission.
Admission Severity of Illness
To account for severity of illness at admission, we used a predicted mortality measure developed at KP.14 This method strictly utilizes information available prior to hospital admissionincluding that from the ED; variables included age, gender, admitting diagnosis, and measures of laboratory test and comorbid disease burden. The method, derived using 259,669 KP hospitalizations, produced a c‐statistic of 0.88 for inpatient mortality; external validation, based on 188,724 hospitalizations in Ottawa, produced a c‐statistic of 0.92.14, 18
Admitting diagnoses were based on admission International Classification of Diseases, 9th revision (ICD‐9) codes, and grouped into 44 broad Primary Conditions based on pathophysiologic plausibility and mortality rates.14 The method also quantified each patient's physiologic derangement and preexisting disease burden based on automated laboratory and comorbidity measuresthe Laboratory Acute Physiology Score (LAPS) and the Comorbidity Point Score (COPS).14
In brief, the LAPS was derived from 14 possible test results obtained in the 24‐hour time period preceding hospitalization, including: anion gap; arterial pH, PaCO2, and PaO2; bicarbonate; serum levels of albumin, total bilirubin, creatinine, glucose, sodium, and troponin I; blood urea nitrogen; creatinine; hematocrit; and total white blood cell count.14 The COPS was calculated from each subject's inpatient and outpatient diagnoses, based on Diagnostic Cost Groups software,19 during the 12‐month period preceding hospitalization.14 Increasing LAPS and COPS values were associated with increases in hospital mortality; detailed information about the development, application, and validation are available in previous work.14, 18
Statistical Analysis
Evaluating excess adverse outcomes associated with unplanned transfers requires adequate control of confounding variables. Our approach to reduce confounding was multivariable case matchinga technique used for assessing treatment effects in observational data.20, 21 Patients with unplanned transfersidentified as caseswere matched with similar controls based on observed variables at the time of hospital admission.
We first matched patients with unplanned ICU transfers within 24 hours of hospital admission to direct ICU admit controls based on predicted in‐hospital mortality (to within 1%); age (by decade); gender; and admitting diagnosis. If a case was matched to multiple controls, we selected 1 control with the most similar admission characteristics (weekday or weekend admission and nursing shift). The risk of death associated with unplanned transfers was estimated using multivariable conditional logistic regression. In secondary analysis, we repeated this analysis only among case‐control pairs within the same hospital facilities.
To cross‐validate the results from multivariable matching techniques, we also performed mixed‐effects multivariable logistic regression including all early unplanned transfer patients and direct ICU admit patients, while adjusting for predicted hospital mortality, age, gender, admitting diagnosis, LAPS, COPS, weekend versus weekday admission, nursing shift, and hospital facility random effects. We repeated these same analyses where cases were defined as patients transferred to the ICU within 48 hours of hospitalization.
Unplanned Transfer Timing
Using bed history data, we identified the elapsed time from admission to unplanned transfer, and categorized patients in increments of elapsed time from admission to unplanned transfer. Time‐to‐unplanned transfer was summarized using Kaplan‐Meier curve.
All analyses were performed in Stata/IC 11.0 for Mac (StataCorp LP, College Station, TX). Continuous variables were reported as mean standard deviation (SD). Cohort comparisons were performed with analysis of variance (ANOVA). Categorical variables were summarized using frequencies and compared with chi‐squared testing. A P value <0.05 was considered statistically significant.
RESULTS
During the study period, 313,797 medical hospitalizations originated in the ED (Table 1). Overall, patients' mean age was 67 18 years; 53.7% were female. Patient characteristics differed significantly based on the need for ICU admission. For example, average LAPS was highest among patients admitted directly to the ICU and lowest among patients who never required ICU care (P < 0.01). Patients with unplanned ICU transfers during hospitalization had longer length of stay and higher hospital mortality than direct ICU admit patients (P < 0.01). Overall, more than 1 in 15 patients experienced an unplanned transfer to the ICU.
Early Delayed ICU Transfer (by Elapsed Time Since Hospital Admission) | ||||
---|---|---|---|---|
Variable | Overall | Within 24 hr | Within 48 hr | Direct ICU Admit |
| ||||
No. (%) | 313,797 | 6,369 (2.0) | 9,816 (3.1) | 29,929 (9.5) |
Age* | 67 18 | 67 16 | 68 16 | 64 17 |
Female* | 169,358 (53.7) | 3,125 (49.1) | 4,882 (49.7) | 14,488 (48.4) |
Weekend admission* | 83,327 (26.6) | 1,783 (28.0) | 2,733 (27.8) | 8,152 (27.2) |
Nursing shift at admission* | ||||
Day (7 AM‐3 PM) | 65,303 (20.8) | 1,335 (21.0) | 2,112 (21.5) | 7,065 (23.6) |
Evening (3 PM‐11 PM) | 155,037 (49.4) | 2,990 (47.0) | 4,691 (47.8) | 13,158 (44.0) |
Night (11 PM‐7 AM) | 93,457 (29.8) | 2,044 (32.1) | 3,013 (30.7) | 9,706 (32.4) |
Initial hospital location* | ||||
Ward | 234,915 (82.8) | 5,177 (81.3) | 7,987 (81.4) | |
Transitional care unit | 48,953 (17.2) | 1,192 (18.7) | 1,829 (18.6) | |
LAPS* | 24 19 | 28 20 | 28 20 | 35 25 |
COPS* | 98 67 | 105 70 | 106 70 | 99 71 |
Length of stay (days) | 4.6 7.5 | 8.4 12.2 | 9.1 13.4 | 6.4 9.5 |
In‐hospital mortality | 12,686 (4.0) | 800 (12.6) | 1,388 (14.1) | 3,602 (12.0) |
The majority of unplanned transfers occurred within the first 48 hours of hospitalization (57.6%, Figure 1); nearly 80% occurred within the first 4 days. The rate of unplanned transfer peaked within 24 hours of hospital admission and decreased gradually as elapsed hospital LOS increased (Figure 1). While most patients experienced a single unplanned ICU transfer, 12.7% required multiple transfers to the ICU throughout their hospitalization.
Multivariable case matching between unplanned transfer cases within 24 hours of admission and direct ICU admit controls resulted in 5839 (92%) case‐control pairs (Table 2). Matched pairs were most frequently admitted with diagnoses in Primary Condition groups that included respiratory infections and pneumonia (15.6%); angina, acute myocardial infarction (AMI), and heart failure (15.6%); or gastrointestinal bleeding (13.8%).
ICU Cohorts (by Elapsed Time to Transfer Since Hospital Admission) | ||||
---|---|---|---|---|
Within 24 hr (n = 5,839) | Within 48 hr (n = 8,976) | |||
Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | |
| ||||
Age | 67 16 | 66 16 | 67 16 | 67 16 |
Female | 2,868 (49.1) | 2,868 (49.1) | 4,477 (49.9) | 4,477 (49.9) |
Admitting diagnosis | ||||
Pneumonia | 911 (15.6) | 911 (15.6) | 1,526 (17.0) | 1,526 (17.0) |
Heart failure or MI | 909 (15.6) | 909 (15.6) | 1,331 (14.8) | 1,331 (14.8) |
Gastrointestinal bleeding | 806 (13.8) | 806 (13.8) | 1,191 (13.3) | 1,191 (13.3) |
Infections (including sepsis) | 295 (5.1) | 295 (5.1) | 474 (5.3) | 474 (5.3) |
Outcomes | ||||
Length of stay (days)* | 8 12 | 6 9 | 9 13 | 6 9 |
In‐hospital mortality* | 678 (11.6) | 498 (8.5) | 1,181 (13.2) | 814 (9.1) |
In‐hospital mortality was significantly higher among cases (11.6%) than among ICU controls (8.5%, P < 0.001); mean LOS was also longer among cases (8 12 days) than among controls (6 9 days, P < 0.001). Unplanned transfer cases were at an increased odds of death when compared with ICU controls (adjusted odds ratio [OR], 1.44; 95% confidence interval [CI], 1.26‐1.64; P < 0.001); they also had a significantly higher observed‐to‐expected mortality ratio. When cases and controls were matched by hospital facility, the number of case‐control pairs decreased (2949 pairs; 42% matching frequency) but the odds of death was of similar magnitude (OR, 1.43; 95% CI, 1.21‐1.68; P < 0.001). Multivariable mixed‐effects logistic regression including all early unplanned transfer and direct ICU admit patients produced an effect size of similar magnitude (OR, 1.37; 95% CI, 1.24‐1.50; P < 0.001).
Results were similar when cases were limited to patients with transfers within 12 hours of admission; mortality was 10.9% among cases and 9.1% among controls (P = 0.02). When including patients with unplanned transfers within 48 hours of hospital admission, the difference in mortality between cases and controls increased (13.2% vs 9.1%, P < 0.001). The odds of death among patients with unplanned transfers increased as the elapsed time between admission and ICU transfer lengthened (Figure 2); the adjusted OR was statistically significant at each point between 8 and 48 hours.
When stratified by admitting diagnosis groups, cases with unplanned transfers within the first 48 hours had increased mortality compared with matched controls in some categories (Table 3). For example, for patients in the respiratory infection and pneumonia group, mortality was 16.8% among unplanned transfer cases and 13.0% among early matched ICU controls (P < 0.01). A similar pattern was present in groups including: gastrointestinal bleeding, chronic obstructive pulmonary disease (COPD) exacerbation, and seizure groups (Table 3). However, for patients with AMI alone, mortality was 5.0% among cases and 3.7% among matched controls (P = 0.12). Patients with sepsis had a mortality rate of 15.2% among cases and 20.8% among matched controls (P = 0.07). Similarly, patients with stroke had a mortality rate of 12.4% among unplanned transfer cases and 11.4% in the matched controls (P = 0.54).
Primary Condition Group | Mortality in ICU Case‐Control Cohorts, No. (%) | |||
---|---|---|---|---|
Within 24 hr | Within 48 hr | |||
Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | |
| ||||
Respiratory infections | 143 (15.7) | 126 (13.8) | 493 (16.8) | 380 (13.0) |
Angina, heart failure, or MI | 60 (6.6) | 41 (4.5) | 324 (7.7) | 152 (3.6) |
Acute MI alone | 16 (5.7) | 17 (6.1) | 82 (5.0) | 61 (3.7) |
Gastrointestinal bleeding | 96 (11.9) | 55 (6.8) | 549 (19.3) | 188 (6.6) |
Infections including sepsis | 20 (9.8) | 52 (11.2) | 228 (14.8) | 220 (14.2) |
Sepsis alone | 32 (18.9) | 31 (18.3) | 123 (15.2) | 168 (20.8) |
COPD exacerbation | 20 (9.8) | 12 (5.9) | 74 (10.8) | 43 (6.3) |
Stroke | 18 (10.2) | 19 (10.8) | 77 (12.4) | 71 (11.4) |
Seizure | 21 (8.6) | 9 (3.7) | 68 (7.1) | 34 (3.6) |
DISCUSSION
This study found that unplanned ICU transfers were common among medical patients, occurring in 5% of all hospitalizations originating in the ED. The majority of unplanned transfers occurred within 48 hours of admission; the rate of ICU transfers peaked within 24 hours after hospitalization. Compared with patients admitted directly from the ED to the ICU, those transferred early after admission had significantly increased mortality; for example, patients transferred within 24 hours were at a 44% increased odds of hospital death. The adverse outcomes associated with unplanned transfers varied considerably by admission diagnosis subgroups.
Our findings confirm previous reports of increased mortality among patients with unplanned ICU transfers. Escarce and Kelley reported that patients admitted to the ICU from non‐ED locationsincluding wards, intermediate care units, and other hospitalswere at an increased risk of hospital death.1 Multiple subsequent studies have confirmed the increased mortality among patients with unplanned transfers.24, 10, 13, 22, 23 We previously evaluated patients who required a transfer to any higher level of care and reported an observed‐to‐expected mortality ratio of 2.93.4
Fewer studies, however, have evaluated the association between the timing of unplanned transfers and inpatient outcomes; previous small reports suggest that delays in ICU transfer adversely affect mortality and length of stay.12, 13, 24 Parkhe et al. compared 99 direct ICU admit patients with 23 who experienced early unplanned transfers; mortality at 30 days was significantly higher among patients with unplanned transfers.13 The current multifacility study included considerably more patients and confirmed an in‐hospital mortality gapalbeit a smaller onebetween patients with early transfers and those directly admitted to the ICU.
We focused on unplanned transfers during the earliest phase of hospitalization to identify patients who might benefit from improved recognition of, and intervention for, impending critical illness. We found that even patients requiring transfers within 8 hours of hospital admission were at an increased risk of death. Bapoje et al. recently reported that as many as 80% of early unplanned transfers were preventable and that most resulted from inappropriate admission triage.11 Together, these findings suggest that heightened attention to identifying such patients at admission or within the first day of hospitalizationwhen the rates of unplanned transfers peakis critical.
Several important limitations should be recognized in interpreting these results. First, this study was not designed to specifically identify the reasons for unplanned transfers, limiting our ability to characterize episodes in which timely care could have prevented excess mortality. Notably, while previous work suggests that many early unplanned transfers might be prevented with appropriate triage, it is likely that some excess deaths are not preventable even if every patient could be admitted to the ICU directly.
We were able to characterize patient outcomes by admitting diagnoses. Patients admitted for pneumonia and respiratory infection, gastrointestinal bleeding, COPD exacerbation, or seizures demonstrated excess mortality compared with matched ICU controls, while those with AMI, sepsis, and stroke did not. It is possible that differences in diagnosis‐specific excess mortality resulted from increasing adherence to well‐defined practice guidelines for specific high‐risk conditions.2527 For example, international awareness campaigns for the treatment of sepsis, AMI, and strokeSurviving Sepsis, Door‐to‐Balloon, and F.A.S.T.emphasize early interventions to minimize morbidity and mortality.
Second, the data utilized in this study were based on automated variables extracted from the electronic medical record. Mortality prediction models based on automated variables have demonstrated excellent performance among ICU and non‐ICU populations14, 18, 28; however, the inclusion of additional data (eg, vital signs or neurological status) would likely improve baseline risk adjustment.5, 10, 2931 Multiple studies have demonstrated that vital signs and clinician judgment can predict patients at an increased risk of deterioration.5, 10, 2931 Such data might also provide insight into residual factors that influenced clinicians' decisions to triage patients to an ICU versus non‐ICU admissiona focus area of our ongoing research efforts. Utilizing electronically available data, however, facilitated the identification of a cohort of patients far larger than that in prior studies. Where previous work has also been limited by substantial variability in baseline characteristics among study subjects,1, 2, 12, 13 our large sample produced a high percentage of multivariable case matches.
Third, we chose to match patients with a severity of illness index based on variables available at the time of hospital admission. While this mortality prediction model has demonstrated excellent performance in internal and external populations,14, 18 it is calibrated for general inpatient, rather than critically ill, populations. It remains possible that case matching with ICU‐specific severity of illness scores might alter matching characteristics, however, previous studies suggest that severity of illness, as measured by these scores, is comparable between direct ICU admits and early ICU transfers.13 Importantly, our matching procedure avoided the potential confounding known to exist with the use of prediction models based on discharge or intra‐hospitalization data.32, 33
Finally, while we were able to evaluate unplanned transfer timing in a multifacility sample, all patient care occurred within a large integrated healthcare delivery system. The overall observed mortality in our study was lower than that reported in prior studies which considered more limited patient cohorts.1, 2, 12, 13, 22 Thus, differences in patient case‐mix or ICU structure must be considered when applying our results to other healthcare delivery systems.
This hypothesis‐generating study, based on a large, multifacility sample of hospitalizations, suggests several areas of future investigation. Future work should detail specific aspects of care among patients with unplanned transfer, including: evaluating the structures and processes involved in triage decisions, measuring the effects on mortality through implementation of interventions (eg, rapid response teams or diagnosis‐specific treatment protocols), and defining the causes and risk factors for unplanned transfers by elapsed time.
In conclusion, the risk of an unplanned ICU transfera common event among hospitalized patientsis highest within 24 hours of hospitalization. Patients with early unplanned transfers have increased mortality and length of stay compared to those admitted directly to the ICU. Even patients transferred to the ICU within 8 hours of hospital admission are at an increased risk of death when compared with those admitted directly. Substantial variability in unplanned transfer outcomes exists based on admitting diagnoses. Future research should characterize unplanned transfers in greater detail with the goal of identifying patients that would benefit from improved triage and early ICU transfer.
- Admission source to the medical intensive care unit predicts hospital death independent of APACHE II score.JAMA.1990;264(18):2389–2394. , .
- Unplanned admission to intensive care after emergency hospitalisation: risk factors and development of a nomogram for individualising risk.Resuscitation.2009;80(2):224–230. , , , , , .
- Outcome of intensive care patients in a group of British intensive care units.Crit Care Med.1998;26(8):1337–1345. , .
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS).J Hosp Med.2010;6(2):74–80. , , , , , .
- Medical patients at high risk for catastrophic deterioration.Crit Care Med.1987;15(5):510–515. , .
- Introduction of the medical emergency team (MET) system: a cluster‐randomised controlled trial.Lancet.2005;365(9477):2091–2097. , , , et al.
- Effect of a rapid response team on hospital‐wide mortality and code rates outside the ICU in a children's hospital.JAMA.2007;298(19):2267–2274. , , , et al.
- Validity of unplanned admission to an intensive care unit as a measure of patient safety in surgical patients.Anesthesiology.2005;103(6):1121–1129. , , , , , .
- The 100,000 lives campaign: setting a goal and a deadline for improving health care quality.JAMA.2006;295(3):324–327. , , , .
- Duration of life‐threatening antecedents prior to intensive care admission.Intensive Care Med.2002;28(11):1629–1634. , , , et al.
- Unplanned transfers to a medical intensive care unit: causes and relationship to preventable errors in care.J Hosp Med.2011;6(2):68–72. , , , .
- Inpatient transfers to the intensive care unit: delays are associated with increased mortality and morbidity.J Gen Intern Med.2003;18(2):77–83. , , , , .
- Outcome of emergency department patients with delayed admission to an intensive care unit.Emerg Med (Fremantle).2002;14(1):50–57. , , , .
- Risk‐adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases.Med Care.2008;46(3):232–239. , , , , , .
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases.Am J Manag Care.2008;14(3):158–166. , , , et al.
- Linking automated databases for research in managed care settings.Ann Intern Med.1997;127(8 pt 2):719–724. .
- Anticoagulation therapy for stroke prevention in atrial fibrillation: how well do randomized trials translate into clinical practice?JAMA.2003;290(20):2685–2692. , , , et al.
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population.J Clin Epidemiol.2009;63(7):798–803. , , , .
- Refinements to the diagnostic cost group (DCG) model.Inquiry.1995;32(4):418–429. , .
- Excess length of stay, charges, and mortality attributable to medical injuries during hospitalization.JAMA.2003;290(14):1868–1874. , .
- Optimal matching in observational studies.J Am Stat Assoc.1989;84:1024–1032. .
- Admissions to intensive care units from emergency departments: a descriptive study.Emerg Med J.2005;22(6):423–428. , , , .
- Using administrative data to develop a nomogram for individualising risk of unplanned admission to intensive care.Resuscitation.2008;79(2):241–248. , , , .
- Unplanned intensive care unit transfers: a useful tool to improve quality of care [abstract]. In: Hospital Medicine 2010 abstract booklet. Society of Hospital Medicine 2010 Annual Meeting, April 9–11, 2010, Washington, DC;2010:10–11. , , , .
- Surviving Sepsis Campaign: international guidelines for management of severe sepsis and septic shock: 2008.Crit Care Med.2008;36(1):296–327. , , , et al.
- 2009 Focused Updates: ACC/AHA Guidelines for the Management of Patients With ST‐Elevation Myocardial Infarction (updating the 2004 Guideline and 2007 Focused Update) and ACC/AHA/SCAI Guidelines on Percutaneous Coronary Intervention (updating the 2005 Guideline and 2007 Focused Update): a report of the American College of Cardiology Foundation/American Heart Association Task Force on Practice Guidelines.Circulation.2009;120(22):2271–2306. , , , et al.
- Translating evidence into practice: a decade of efforts by the American Heart Association/American Stroke Association to reduce death and disability due to stroke: a presidential advisory from the American Heart Association/American Stroke Association.Stroke.2010;41(5):1051–1065. , , , et al.
- Veterans Affairs intensive care unit risk adjustment model: validation, updating, recalibration.Crit Care Med.2008;36(4):1031–1042. , , , et al.
- Recommended guidelines for monitoring, reporting, and conducting research on medical emergency team, outreach, and rapid response systems: an Utstein‐style scientific statement: a scientific statement from the International Liaison Committee on Resuscitation (American Heart Association, Australian Resuscitation Council, European Resuscitation Council, Heart and Stroke Foundation of Canada, InterAmerican Heart Foundation, Resuscitation Council of Southern Africa, and the New Zealand Resuscitation Council); the American Heart Association Emergency Cardiovascular Care Committee; the Council on Cardiopulmonary, Perioperative, and Critical Care; and the Interdisciplinary Working Group on Quality of Care and Outcomes Research.Circulation.2007;116(21):2481–2500. , , , et al.
- Realizing the potential of clinical judgment: a real‐time strategy for predicting outcomes and cost for medical inpatients.Am J Med.2000;109(3):189–195. , , , , , .
- Physiological values and procedures in the 24 h before ICU admission from the ward.Anaesthesia.1999;54(6):529–534. , , .
- Predicting who dies depends on how severity is measured: implications for evaluating patient outcomes.Ann Intern Med.1995;123(10):763–770. , , , , , .
- Enhancement of claims data to improve risk adjustment of hospital mortality.JAMA.2007;297(1):71–76. , , , et al.
Hospitalized patients who require transfer from medical wards to the intensive care unit (ICU) have high in‐hospital mortality, in some reports exceeding 55%.14 In a previous report in this journal, we found that while these unplanned ICU transfers occurred in only 4% of hospitalizations, they were present in nearly one‐quarter of fatal hospitalizations and were associated with substantial increases in resource utilization.4 For these reasons, interventions aimed at identifying and treating this high‐risk group have received considerable attention and have been proposed as measures of inpatient safety.2, 49
Notably, mortality among patients with unplanned ICU transfers exceeds mortality among patients admitted to the ICU directly from the emergency department (ED)a group traditionally considered to have the highest risk of death.13, 10 Previous single‐center studies suggest that increased mortality rates are present even among patients transferred within 24 hours of hospital admission, and reinforce the notion that earlier recognition of critical illness may result in improved outcomes.1113 However, these studies have been performed primarily in small cohorts of heterogeneous patients, and may obscure the independent effect of unplanned transfers on mortality and hamper efforts to use unplanned transfer rates as a metric of healthcare quality.1, 2, 4, 9
In this study, we evaluated early unplanned ICU transfers drawn from a cohort of 499,995 hospitalizations in an integrated healthcare delivery system. Using patient data, extracted from the automated electronic medical record, we matched unplanned transfer cases to patients directly admitted to the ICU and described the association between delayed ICU transfers and adverse outcomes.
METHODS
Setting and Participants
We performed a retrospective analysis of adult patient (age 18 years) hospitalizations at 21 Northern California Kaiser Permanente (KP) Medical Care Program hospitals between January 2007 and December 2009. This work expanded on our previous report of hospital stays from November 2006 to January 2008.4 The 21 study hospitals used the same electronic health information systems; databases captured admission, discharge, and bed history data. The use of these databases for research has been described in our previous study and other reports; hospital characteristics, unit staffing, and resource levels have also been detailed previously.4, 1417 This study was approved by the KP Institutional Review Board.
Identifying Unplanned Transfers
We evaluated patients with medical hospitalizationsdefined as those whose first hospital location was not in a surgical setting such as the operating room or post‐anesthesia recovery areawhose admission originated in the ED; patients admitted for surgery were removed because of significant differences in observed mortality (see Supporting Information Appendix Figure 1 and Appendix Table 1 in the online version of this article). Patients whose admission did not originate in the ED were excluded to eliminate confounding resulting from differences in preadmission care. We also excluded patients admitted for gynecological and pregnancy‐related care because of low hospital mortality.
Initial patient locations included the medical wards (wards); the transitional care unit (TCU); and the intensive care unit (ICU). Bed history data, based on time stamps and available for all patients, were used to track patient locations from the time of admission, defined as the first non‐ED hospital location, until discharge. Patient length of stay (LOS) was calculated at each location and for the entire hospitalization.
Transfers to the ICU after a patient's initial admission to the ward or TCU were termed unplanned (or delayed) ICU transfers; patients admitted from the ED to the ICU were termed direct ICU admit patients. Direct ICU admit patients were excluded from the unplanned transfer group even if they required a readmission to the ICU later in their hospital course. We focused on patients with unplanned ICU transfers early after hospitalization to identify those in whom prompt recognition and intervention could be effective; thus, our primary analyses were on patients with transfers within 24 hours of admission. In secondary analysis, we also evaluated patients with unplanned ICU transfers occurring within 48 hours after hospital admission.
Admission Severity of Illness
To account for severity of illness at admission, we used a predicted mortality measure developed at KP.14 This method strictly utilizes information available prior to hospital admissionincluding that from the ED; variables included age, gender, admitting diagnosis, and measures of laboratory test and comorbid disease burden. The method, derived using 259,669 KP hospitalizations, produced a c‐statistic of 0.88 for inpatient mortality; external validation, based on 188,724 hospitalizations in Ottawa, produced a c‐statistic of 0.92.14, 18
Admitting diagnoses were based on admission International Classification of Diseases, 9th revision (ICD‐9) codes, and grouped into 44 broad Primary Conditions based on pathophysiologic plausibility and mortality rates.14 The method also quantified each patient's physiologic derangement and preexisting disease burden based on automated laboratory and comorbidity measuresthe Laboratory Acute Physiology Score (LAPS) and the Comorbidity Point Score (COPS).14
In brief, the LAPS was derived from 14 possible test results obtained in the 24‐hour time period preceding hospitalization, including: anion gap; arterial pH, PaCO2, and PaO2; bicarbonate; serum levels of albumin, total bilirubin, creatinine, glucose, sodium, and troponin I; blood urea nitrogen; creatinine; hematocrit; and total white blood cell count.14 The COPS was calculated from each subject's inpatient and outpatient diagnoses, based on Diagnostic Cost Groups software,19 during the 12‐month period preceding hospitalization.14 Increasing LAPS and COPS values were associated with increases in hospital mortality; detailed information about the development, application, and validation are available in previous work.14, 18
Statistical Analysis
Evaluating excess adverse outcomes associated with unplanned transfers requires adequate control of confounding variables. Our approach to reduce confounding was multivariable case matchinga technique used for assessing treatment effects in observational data.20, 21 Patients with unplanned transfersidentified as caseswere matched with similar controls based on observed variables at the time of hospital admission.
We first matched patients with unplanned ICU transfers within 24 hours of hospital admission to direct ICU admit controls based on predicted in‐hospital mortality (to within 1%); age (by decade); gender; and admitting diagnosis. If a case was matched to multiple controls, we selected 1 control with the most similar admission characteristics (weekday or weekend admission and nursing shift). The risk of death associated with unplanned transfers was estimated using multivariable conditional logistic regression. In secondary analysis, we repeated this analysis only among case‐control pairs within the same hospital facilities.
To cross‐validate the results from multivariable matching techniques, we also performed mixed‐effects multivariable logistic regression including all early unplanned transfer patients and direct ICU admit patients, while adjusting for predicted hospital mortality, age, gender, admitting diagnosis, LAPS, COPS, weekend versus weekday admission, nursing shift, and hospital facility random effects. We repeated these same analyses where cases were defined as patients transferred to the ICU within 48 hours of hospitalization.
Unplanned Transfer Timing
Using bed history data, we identified the elapsed time from admission to unplanned transfer, and categorized patients in increments of elapsed time from admission to unplanned transfer. Time‐to‐unplanned transfer was summarized using Kaplan‐Meier curve.
All analyses were performed in Stata/IC 11.0 for Mac (StataCorp LP, College Station, TX). Continuous variables were reported as mean standard deviation (SD). Cohort comparisons were performed with analysis of variance (ANOVA). Categorical variables were summarized using frequencies and compared with chi‐squared testing. A P value <0.05 was considered statistically significant.
RESULTS
During the study period, 313,797 medical hospitalizations originated in the ED (Table 1). Overall, patients' mean age was 67 18 years; 53.7% were female. Patient characteristics differed significantly based on the need for ICU admission. For example, average LAPS was highest among patients admitted directly to the ICU and lowest among patients who never required ICU care (P < 0.01). Patients with unplanned ICU transfers during hospitalization had longer length of stay and higher hospital mortality than direct ICU admit patients (P < 0.01). Overall, more than 1 in 15 patients experienced an unplanned transfer to the ICU.
Early Delayed ICU Transfer (by Elapsed Time Since Hospital Admission) | ||||
---|---|---|---|---|
Variable | Overall | Within 24 hr | Within 48 hr | Direct ICU Admit |
| ||||
No. (%) | 313,797 | 6,369 (2.0) | 9,816 (3.1) | 29,929 (9.5) |
Age* | 67 18 | 67 16 | 68 16 | 64 17 |
Female* | 169,358 (53.7) | 3,125 (49.1) | 4,882 (49.7) | 14,488 (48.4) |
Weekend admission* | 83,327 (26.6) | 1,783 (28.0) | 2,733 (27.8) | 8,152 (27.2) |
Nursing shift at admission* | ||||
Day (7 AM‐3 PM) | 65,303 (20.8) | 1,335 (21.0) | 2,112 (21.5) | 7,065 (23.6) |
Evening (3 PM‐11 PM) | 155,037 (49.4) | 2,990 (47.0) | 4,691 (47.8) | 13,158 (44.0) |
Night (11 PM‐7 AM) | 93,457 (29.8) | 2,044 (32.1) | 3,013 (30.7) | 9,706 (32.4) |
Initial hospital location* | ||||
Ward | 234,915 (82.8) | 5,177 (81.3) | 7,987 (81.4) | |
Transitional care unit | 48,953 (17.2) | 1,192 (18.7) | 1,829 (18.6) | |
LAPS* | 24 19 | 28 20 | 28 20 | 35 25 |
COPS* | 98 67 | 105 70 | 106 70 | 99 71 |
Length of stay (days) | 4.6 7.5 | 8.4 12.2 | 9.1 13.4 | 6.4 9.5 |
In‐hospital mortality | 12,686 (4.0) | 800 (12.6) | 1,388 (14.1) | 3,602 (12.0) |
The majority of unplanned transfers occurred within the first 48 hours of hospitalization (57.6%, Figure 1); nearly 80% occurred within the first 4 days. The rate of unplanned transfer peaked within 24 hours of hospital admission and decreased gradually as elapsed hospital LOS increased (Figure 1). While most patients experienced a single unplanned ICU transfer, 12.7% required multiple transfers to the ICU throughout their hospitalization.
Multivariable case matching between unplanned transfer cases within 24 hours of admission and direct ICU admit controls resulted in 5839 (92%) case‐control pairs (Table 2). Matched pairs were most frequently admitted with diagnoses in Primary Condition groups that included respiratory infections and pneumonia (15.6%); angina, acute myocardial infarction (AMI), and heart failure (15.6%); or gastrointestinal bleeding (13.8%).
ICU Cohorts (by Elapsed Time to Transfer Since Hospital Admission) | ||||
---|---|---|---|---|
Within 24 hr (n = 5,839) | Within 48 hr (n = 8,976) | |||
Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | |
| ||||
Age | 67 16 | 66 16 | 67 16 | 67 16 |
Female | 2,868 (49.1) | 2,868 (49.1) | 4,477 (49.9) | 4,477 (49.9) |
Admitting diagnosis | ||||
Pneumonia | 911 (15.6) | 911 (15.6) | 1,526 (17.0) | 1,526 (17.0) |
Heart failure or MI | 909 (15.6) | 909 (15.6) | 1,331 (14.8) | 1,331 (14.8) |
Gastrointestinal bleeding | 806 (13.8) | 806 (13.8) | 1,191 (13.3) | 1,191 (13.3) |
Infections (including sepsis) | 295 (5.1) | 295 (5.1) | 474 (5.3) | 474 (5.3) |
Outcomes | ||||
Length of stay (days)* | 8 12 | 6 9 | 9 13 | 6 9 |
In‐hospital mortality* | 678 (11.6) | 498 (8.5) | 1,181 (13.2) | 814 (9.1) |
In‐hospital mortality was significantly higher among cases (11.6%) than among ICU controls (8.5%, P < 0.001); mean LOS was also longer among cases (8 12 days) than among controls (6 9 days, P < 0.001). Unplanned transfer cases were at an increased odds of death when compared with ICU controls (adjusted odds ratio [OR], 1.44; 95% confidence interval [CI], 1.26‐1.64; P < 0.001); they also had a significantly higher observed‐to‐expected mortality ratio. When cases and controls were matched by hospital facility, the number of case‐control pairs decreased (2949 pairs; 42% matching frequency) but the odds of death was of similar magnitude (OR, 1.43; 95% CI, 1.21‐1.68; P < 0.001). Multivariable mixed‐effects logistic regression including all early unplanned transfer and direct ICU admit patients produced an effect size of similar magnitude (OR, 1.37; 95% CI, 1.24‐1.50; P < 0.001).
Results were similar when cases were limited to patients with transfers within 12 hours of admission; mortality was 10.9% among cases and 9.1% among controls (P = 0.02). When including patients with unplanned transfers within 48 hours of hospital admission, the difference in mortality between cases and controls increased (13.2% vs 9.1%, P < 0.001). The odds of death among patients with unplanned transfers increased as the elapsed time between admission and ICU transfer lengthened (Figure 2); the adjusted OR was statistically significant at each point between 8 and 48 hours.
When stratified by admitting diagnosis groups, cases with unplanned transfers within the first 48 hours had increased mortality compared with matched controls in some categories (Table 3). For example, for patients in the respiratory infection and pneumonia group, mortality was 16.8% among unplanned transfer cases and 13.0% among early matched ICU controls (P < 0.01). A similar pattern was present in groups including: gastrointestinal bleeding, chronic obstructive pulmonary disease (COPD) exacerbation, and seizure groups (Table 3). However, for patients with AMI alone, mortality was 5.0% among cases and 3.7% among matched controls (P = 0.12). Patients with sepsis had a mortality rate of 15.2% among cases and 20.8% among matched controls (P = 0.07). Similarly, patients with stroke had a mortality rate of 12.4% among unplanned transfer cases and 11.4% in the matched controls (P = 0.54).
Primary Condition Group | Mortality in ICU Case‐Control Cohorts, No. (%) | |||
---|---|---|---|---|
Within 24 hr | Within 48 hr | |||
Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | |
| ||||
Respiratory infections | 143 (15.7) | 126 (13.8) | 493 (16.8) | 380 (13.0) |
Angina, heart failure, or MI | 60 (6.6) | 41 (4.5) | 324 (7.7) | 152 (3.6) |
Acute MI alone | 16 (5.7) | 17 (6.1) | 82 (5.0) | 61 (3.7) |
Gastrointestinal bleeding | 96 (11.9) | 55 (6.8) | 549 (19.3) | 188 (6.6) |
Infections including sepsis | 20 (9.8) | 52 (11.2) | 228 (14.8) | 220 (14.2) |
Sepsis alone | 32 (18.9) | 31 (18.3) | 123 (15.2) | 168 (20.8) |
COPD exacerbation | 20 (9.8) | 12 (5.9) | 74 (10.8) | 43 (6.3) |
Stroke | 18 (10.2) | 19 (10.8) | 77 (12.4) | 71 (11.4) |
Seizure | 21 (8.6) | 9 (3.7) | 68 (7.1) | 34 (3.6) |
DISCUSSION
This study found that unplanned ICU transfers were common among medical patients, occurring in 5% of all hospitalizations originating in the ED. The majority of unplanned transfers occurred within 48 hours of admission; the rate of ICU transfers peaked within 24 hours after hospitalization. Compared with patients admitted directly from the ED to the ICU, those transferred early after admission had significantly increased mortality; for example, patients transferred within 24 hours were at a 44% increased odds of hospital death. The adverse outcomes associated with unplanned transfers varied considerably by admission diagnosis subgroups.
Our findings confirm previous reports of increased mortality among patients with unplanned ICU transfers. Escarce and Kelley reported that patients admitted to the ICU from non‐ED locationsincluding wards, intermediate care units, and other hospitalswere at an increased risk of hospital death.1 Multiple subsequent studies have confirmed the increased mortality among patients with unplanned transfers.24, 10, 13, 22, 23 We previously evaluated patients who required a transfer to any higher level of care and reported an observed‐to‐expected mortality ratio of 2.93.4
Fewer studies, however, have evaluated the association between the timing of unplanned transfers and inpatient outcomes; previous small reports suggest that delays in ICU transfer adversely affect mortality and length of stay.12, 13, 24 Parkhe et al. compared 99 direct ICU admit patients with 23 who experienced early unplanned transfers; mortality at 30 days was significantly higher among patients with unplanned transfers.13 The current multifacility study included considerably more patients and confirmed an in‐hospital mortality gapalbeit a smaller onebetween patients with early transfers and those directly admitted to the ICU.
We focused on unplanned transfers during the earliest phase of hospitalization to identify patients who might benefit from improved recognition of, and intervention for, impending critical illness. We found that even patients requiring transfers within 8 hours of hospital admission were at an increased risk of death. Bapoje et al. recently reported that as many as 80% of early unplanned transfers were preventable and that most resulted from inappropriate admission triage.11 Together, these findings suggest that heightened attention to identifying such patients at admission or within the first day of hospitalizationwhen the rates of unplanned transfers peakis critical.
Several important limitations should be recognized in interpreting these results. First, this study was not designed to specifically identify the reasons for unplanned transfers, limiting our ability to characterize episodes in which timely care could have prevented excess mortality. Notably, while previous work suggests that many early unplanned transfers might be prevented with appropriate triage, it is likely that some excess deaths are not preventable even if every patient could be admitted to the ICU directly.
We were able to characterize patient outcomes by admitting diagnoses. Patients admitted for pneumonia and respiratory infection, gastrointestinal bleeding, COPD exacerbation, or seizures demonstrated excess mortality compared with matched ICU controls, while those with AMI, sepsis, and stroke did not. It is possible that differences in diagnosis‐specific excess mortality resulted from increasing adherence to well‐defined practice guidelines for specific high‐risk conditions.2527 For example, international awareness campaigns for the treatment of sepsis, AMI, and strokeSurviving Sepsis, Door‐to‐Balloon, and F.A.S.T.emphasize early interventions to minimize morbidity and mortality.
Second, the data utilized in this study were based on automated variables extracted from the electronic medical record. Mortality prediction models based on automated variables have demonstrated excellent performance among ICU and non‐ICU populations14, 18, 28; however, the inclusion of additional data (eg, vital signs or neurological status) would likely improve baseline risk adjustment.5, 10, 2931 Multiple studies have demonstrated that vital signs and clinician judgment can predict patients at an increased risk of deterioration.5, 10, 2931 Such data might also provide insight into residual factors that influenced clinicians' decisions to triage patients to an ICU versus non‐ICU admissiona focus area of our ongoing research efforts. Utilizing electronically available data, however, facilitated the identification of a cohort of patients far larger than that in prior studies. Where previous work has also been limited by substantial variability in baseline characteristics among study subjects,1, 2, 12, 13 our large sample produced a high percentage of multivariable case matches.
Third, we chose to match patients with a severity of illness index based on variables available at the time of hospital admission. While this mortality prediction model has demonstrated excellent performance in internal and external populations,14, 18 it is calibrated for general inpatient, rather than critically ill, populations. It remains possible that case matching with ICU‐specific severity of illness scores might alter matching characteristics, however, previous studies suggest that severity of illness, as measured by these scores, is comparable between direct ICU admits and early ICU transfers.13 Importantly, our matching procedure avoided the potential confounding known to exist with the use of prediction models based on discharge or intra‐hospitalization data.32, 33
Finally, while we were able to evaluate unplanned transfer timing in a multifacility sample, all patient care occurred within a large integrated healthcare delivery system. The overall observed mortality in our study was lower than that reported in prior studies which considered more limited patient cohorts.1, 2, 12, 13, 22 Thus, differences in patient case‐mix or ICU structure must be considered when applying our results to other healthcare delivery systems.
This hypothesis‐generating study, based on a large, multifacility sample of hospitalizations, suggests several areas of future investigation. Future work should detail specific aspects of care among patients with unplanned transfer, including: evaluating the structures and processes involved in triage decisions, measuring the effects on mortality through implementation of interventions (eg, rapid response teams or diagnosis‐specific treatment protocols), and defining the causes and risk factors for unplanned transfers by elapsed time.
In conclusion, the risk of an unplanned ICU transfera common event among hospitalized patientsis highest within 24 hours of hospitalization. Patients with early unplanned transfers have increased mortality and length of stay compared to those admitted directly to the ICU. Even patients transferred to the ICU within 8 hours of hospital admission are at an increased risk of death when compared with those admitted directly. Substantial variability in unplanned transfer outcomes exists based on admitting diagnoses. Future research should characterize unplanned transfers in greater detail with the goal of identifying patients that would benefit from improved triage and early ICU transfer.
Hospitalized patients who require transfer from medical wards to the intensive care unit (ICU) have high in‐hospital mortality, in some reports exceeding 55%.14 In a previous report in this journal, we found that while these unplanned ICU transfers occurred in only 4% of hospitalizations, they were present in nearly one‐quarter of fatal hospitalizations and were associated with substantial increases in resource utilization.4 For these reasons, interventions aimed at identifying and treating this high‐risk group have received considerable attention and have been proposed as measures of inpatient safety.2, 49
Notably, mortality among patients with unplanned ICU transfers exceeds mortality among patients admitted to the ICU directly from the emergency department (ED)a group traditionally considered to have the highest risk of death.13, 10 Previous single‐center studies suggest that increased mortality rates are present even among patients transferred within 24 hours of hospital admission, and reinforce the notion that earlier recognition of critical illness may result in improved outcomes.1113 However, these studies have been performed primarily in small cohorts of heterogeneous patients, and may obscure the independent effect of unplanned transfers on mortality and hamper efforts to use unplanned transfer rates as a metric of healthcare quality.1, 2, 4, 9
In this study, we evaluated early unplanned ICU transfers drawn from a cohort of 499,995 hospitalizations in an integrated healthcare delivery system. Using patient data, extracted from the automated electronic medical record, we matched unplanned transfer cases to patients directly admitted to the ICU and described the association between delayed ICU transfers and adverse outcomes.
METHODS
Setting and Participants
We performed a retrospective analysis of adult patient (age 18 years) hospitalizations at 21 Northern California Kaiser Permanente (KP) Medical Care Program hospitals between January 2007 and December 2009. This work expanded on our previous report of hospital stays from November 2006 to January 2008.4 The 21 study hospitals used the same electronic health information systems; databases captured admission, discharge, and bed history data. The use of these databases for research has been described in our previous study and other reports; hospital characteristics, unit staffing, and resource levels have also been detailed previously.4, 1417 This study was approved by the KP Institutional Review Board.
Identifying Unplanned Transfers
We evaluated patients with medical hospitalizationsdefined as those whose first hospital location was not in a surgical setting such as the operating room or post‐anesthesia recovery areawhose admission originated in the ED; patients admitted for surgery were removed because of significant differences in observed mortality (see Supporting Information Appendix Figure 1 and Appendix Table 1 in the online version of this article). Patients whose admission did not originate in the ED were excluded to eliminate confounding resulting from differences in preadmission care. We also excluded patients admitted for gynecological and pregnancy‐related care because of low hospital mortality.
Initial patient locations included the medical wards (wards); the transitional care unit (TCU); and the intensive care unit (ICU). Bed history data, based on time stamps and available for all patients, were used to track patient locations from the time of admission, defined as the first non‐ED hospital location, until discharge. Patient length of stay (LOS) was calculated at each location and for the entire hospitalization.
Transfers to the ICU after a patient's initial admission to the ward or TCU were termed unplanned (or delayed) ICU transfers; patients admitted from the ED to the ICU were termed direct ICU admit patients. Direct ICU admit patients were excluded from the unplanned transfer group even if they required a readmission to the ICU later in their hospital course. We focused on patients with unplanned ICU transfers early after hospitalization to identify those in whom prompt recognition and intervention could be effective; thus, our primary analyses were on patients with transfers within 24 hours of admission. In secondary analysis, we also evaluated patients with unplanned ICU transfers occurring within 48 hours after hospital admission.
Admission Severity of Illness
To account for severity of illness at admission, we used a predicted mortality measure developed at KP.14 This method strictly utilizes information available prior to hospital admissionincluding that from the ED; variables included age, gender, admitting diagnosis, and measures of laboratory test and comorbid disease burden. The method, derived using 259,669 KP hospitalizations, produced a c‐statistic of 0.88 for inpatient mortality; external validation, based on 188,724 hospitalizations in Ottawa, produced a c‐statistic of 0.92.14, 18
Admitting diagnoses were based on admission International Classification of Diseases, 9th revision (ICD‐9) codes, and grouped into 44 broad Primary Conditions based on pathophysiologic plausibility and mortality rates.14 The method also quantified each patient's physiologic derangement and preexisting disease burden based on automated laboratory and comorbidity measuresthe Laboratory Acute Physiology Score (LAPS) and the Comorbidity Point Score (COPS).14
In brief, the LAPS was derived from 14 possible test results obtained in the 24‐hour time period preceding hospitalization, including: anion gap; arterial pH, PaCO2, and PaO2; bicarbonate; serum levels of albumin, total bilirubin, creatinine, glucose, sodium, and troponin I; blood urea nitrogen; creatinine; hematocrit; and total white blood cell count.14 The COPS was calculated from each subject's inpatient and outpatient diagnoses, based on Diagnostic Cost Groups software,19 during the 12‐month period preceding hospitalization.14 Increasing LAPS and COPS values were associated with increases in hospital mortality; detailed information about the development, application, and validation are available in previous work.14, 18
Statistical Analysis
Evaluating excess adverse outcomes associated with unplanned transfers requires adequate control of confounding variables. Our approach to reduce confounding was multivariable case matchinga technique used for assessing treatment effects in observational data.20, 21 Patients with unplanned transfersidentified as caseswere matched with similar controls based on observed variables at the time of hospital admission.
We first matched patients with unplanned ICU transfers within 24 hours of hospital admission to direct ICU admit controls based on predicted in‐hospital mortality (to within 1%); age (by decade); gender; and admitting diagnosis. If a case was matched to multiple controls, we selected 1 control with the most similar admission characteristics (weekday or weekend admission and nursing shift). The risk of death associated with unplanned transfers was estimated using multivariable conditional logistic regression. In secondary analysis, we repeated this analysis only among case‐control pairs within the same hospital facilities.
To cross‐validate the results from multivariable matching techniques, we also performed mixed‐effects multivariable logistic regression including all early unplanned transfer patients and direct ICU admit patients, while adjusting for predicted hospital mortality, age, gender, admitting diagnosis, LAPS, COPS, weekend versus weekday admission, nursing shift, and hospital facility random effects. We repeated these same analyses where cases were defined as patients transferred to the ICU within 48 hours of hospitalization.
Unplanned Transfer Timing
Using bed history data, we identified the elapsed time from admission to unplanned transfer, and categorized patients in increments of elapsed time from admission to unplanned transfer. Time‐to‐unplanned transfer was summarized using Kaplan‐Meier curve.
All analyses were performed in Stata/IC 11.0 for Mac (StataCorp LP, College Station, TX). Continuous variables were reported as mean standard deviation (SD). Cohort comparisons were performed with analysis of variance (ANOVA). Categorical variables were summarized using frequencies and compared with chi‐squared testing. A P value <0.05 was considered statistically significant.
RESULTS
During the study period, 313,797 medical hospitalizations originated in the ED (Table 1). Overall, patients' mean age was 67 18 years; 53.7% were female. Patient characteristics differed significantly based on the need for ICU admission. For example, average LAPS was highest among patients admitted directly to the ICU and lowest among patients who never required ICU care (P < 0.01). Patients with unplanned ICU transfers during hospitalization had longer length of stay and higher hospital mortality than direct ICU admit patients (P < 0.01). Overall, more than 1 in 15 patients experienced an unplanned transfer to the ICU.
Early Delayed ICU Transfer (by Elapsed Time Since Hospital Admission) | ||||
---|---|---|---|---|
Variable | Overall | Within 24 hr | Within 48 hr | Direct ICU Admit |
| ||||
No. (%) | 313,797 | 6,369 (2.0) | 9,816 (3.1) | 29,929 (9.5) |
Age* | 67 18 | 67 16 | 68 16 | 64 17 |
Female* | 169,358 (53.7) | 3,125 (49.1) | 4,882 (49.7) | 14,488 (48.4) |
Weekend admission* | 83,327 (26.6) | 1,783 (28.0) | 2,733 (27.8) | 8,152 (27.2) |
Nursing shift at admission* | ||||
Day (7 AM‐3 PM) | 65,303 (20.8) | 1,335 (21.0) | 2,112 (21.5) | 7,065 (23.6) |
Evening (3 PM‐11 PM) | 155,037 (49.4) | 2,990 (47.0) | 4,691 (47.8) | 13,158 (44.0) |
Night (11 PM‐7 AM) | 93,457 (29.8) | 2,044 (32.1) | 3,013 (30.7) | 9,706 (32.4) |
Initial hospital location* | ||||
Ward | 234,915 (82.8) | 5,177 (81.3) | 7,987 (81.4) | |
Transitional care unit | 48,953 (17.2) | 1,192 (18.7) | 1,829 (18.6) | |
LAPS* | 24 19 | 28 20 | 28 20 | 35 25 |
COPS* | 98 67 | 105 70 | 106 70 | 99 71 |
Length of stay (days) | 4.6 7.5 | 8.4 12.2 | 9.1 13.4 | 6.4 9.5 |
In‐hospital mortality | 12,686 (4.0) | 800 (12.6) | 1,388 (14.1) | 3,602 (12.0) |
The majority of unplanned transfers occurred within the first 48 hours of hospitalization (57.6%, Figure 1); nearly 80% occurred within the first 4 days. The rate of unplanned transfer peaked within 24 hours of hospital admission and decreased gradually as elapsed hospital LOS increased (Figure 1). While most patients experienced a single unplanned ICU transfer, 12.7% required multiple transfers to the ICU throughout their hospitalization.
Multivariable case matching between unplanned transfer cases within 24 hours of admission and direct ICU admit controls resulted in 5839 (92%) case‐control pairs (Table 2). Matched pairs were most frequently admitted with diagnoses in Primary Condition groups that included respiratory infections and pneumonia (15.6%); angina, acute myocardial infarction (AMI), and heart failure (15.6%); or gastrointestinal bleeding (13.8%).
ICU Cohorts (by Elapsed Time to Transfer Since Hospital Admission) | ||||
---|---|---|---|---|
Within 24 hr (n = 5,839) | Within 48 hr (n = 8,976) | |||
Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | |
| ||||
Age | 67 16 | 66 16 | 67 16 | 67 16 |
Female | 2,868 (49.1) | 2,868 (49.1) | 4,477 (49.9) | 4,477 (49.9) |
Admitting diagnosis | ||||
Pneumonia | 911 (15.6) | 911 (15.6) | 1,526 (17.0) | 1,526 (17.0) |
Heart failure or MI | 909 (15.6) | 909 (15.6) | 1,331 (14.8) | 1,331 (14.8) |
Gastrointestinal bleeding | 806 (13.8) | 806 (13.8) | 1,191 (13.3) | 1,191 (13.3) |
Infections (including sepsis) | 295 (5.1) | 295 (5.1) | 474 (5.3) | 474 (5.3) |
Outcomes | ||||
Length of stay (days)* | 8 12 | 6 9 | 9 13 | 6 9 |
In‐hospital mortality* | 678 (11.6) | 498 (8.5) | 1,181 (13.2) | 814 (9.1) |
In‐hospital mortality was significantly higher among cases (11.6%) than among ICU controls (8.5%, P < 0.001); mean LOS was also longer among cases (8 12 days) than among controls (6 9 days, P < 0.001). Unplanned transfer cases were at an increased odds of death when compared with ICU controls (adjusted odds ratio [OR], 1.44; 95% confidence interval [CI], 1.26‐1.64; P < 0.001); they also had a significantly higher observed‐to‐expected mortality ratio. When cases and controls were matched by hospital facility, the number of case‐control pairs decreased (2949 pairs; 42% matching frequency) but the odds of death was of similar magnitude (OR, 1.43; 95% CI, 1.21‐1.68; P < 0.001). Multivariable mixed‐effects logistic regression including all early unplanned transfer and direct ICU admit patients produced an effect size of similar magnitude (OR, 1.37; 95% CI, 1.24‐1.50; P < 0.001).
Results were similar when cases were limited to patients with transfers within 12 hours of admission; mortality was 10.9% among cases and 9.1% among controls (P = 0.02). When including patients with unplanned transfers within 48 hours of hospital admission, the difference in mortality between cases and controls increased (13.2% vs 9.1%, P < 0.001). The odds of death among patients with unplanned transfers increased as the elapsed time between admission and ICU transfer lengthened (Figure 2); the adjusted OR was statistically significant at each point between 8 and 48 hours.
When stratified by admitting diagnosis groups, cases with unplanned transfers within the first 48 hours had increased mortality compared with matched controls in some categories (Table 3). For example, for patients in the respiratory infection and pneumonia group, mortality was 16.8% among unplanned transfer cases and 13.0% among early matched ICU controls (P < 0.01). A similar pattern was present in groups including: gastrointestinal bleeding, chronic obstructive pulmonary disease (COPD) exacerbation, and seizure groups (Table 3). However, for patients with AMI alone, mortality was 5.0% among cases and 3.7% among matched controls (P = 0.12). Patients with sepsis had a mortality rate of 15.2% among cases and 20.8% among matched controls (P = 0.07). Similarly, patients with stroke had a mortality rate of 12.4% among unplanned transfer cases and 11.4% in the matched controls (P = 0.54).
Primary Condition Group | Mortality in ICU Case‐Control Cohorts, No. (%) | |||
---|---|---|---|---|
Within 24 hr | Within 48 hr | |||
Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | Delayed ICU Transfer (Case) | Direct ICU Admit (Control) | |
| ||||
Respiratory infections | 143 (15.7) | 126 (13.8) | 493 (16.8) | 380 (13.0) |
Angina, heart failure, or MI | 60 (6.6) | 41 (4.5) | 324 (7.7) | 152 (3.6) |
Acute MI alone | 16 (5.7) | 17 (6.1) | 82 (5.0) | 61 (3.7) |
Gastrointestinal bleeding | 96 (11.9) | 55 (6.8) | 549 (19.3) | 188 (6.6) |
Infections including sepsis | 20 (9.8) | 52 (11.2) | 228 (14.8) | 220 (14.2) |
Sepsis alone | 32 (18.9) | 31 (18.3) | 123 (15.2) | 168 (20.8) |
COPD exacerbation | 20 (9.8) | 12 (5.9) | 74 (10.8) | 43 (6.3) |
Stroke | 18 (10.2) | 19 (10.8) | 77 (12.4) | 71 (11.4) |
Seizure | 21 (8.6) | 9 (3.7) | 68 (7.1) | 34 (3.6) |
DISCUSSION
This study found that unplanned ICU transfers were common among medical patients, occurring in 5% of all hospitalizations originating in the ED. The majority of unplanned transfers occurred within 48 hours of admission; the rate of ICU transfers peaked within 24 hours after hospitalization. Compared with patients admitted directly from the ED to the ICU, those transferred early after admission had significantly increased mortality; for example, patients transferred within 24 hours were at a 44% increased odds of hospital death. The adverse outcomes associated with unplanned transfers varied considerably by admission diagnosis subgroups.
Our findings confirm previous reports of increased mortality among patients with unplanned ICU transfers. Escarce and Kelley reported that patients admitted to the ICU from non‐ED locationsincluding wards, intermediate care units, and other hospitalswere at an increased risk of hospital death.1 Multiple subsequent studies have confirmed the increased mortality among patients with unplanned transfers.24, 10, 13, 22, 23 We previously evaluated patients who required a transfer to any higher level of care and reported an observed‐to‐expected mortality ratio of 2.93.4
Fewer studies, however, have evaluated the association between the timing of unplanned transfers and inpatient outcomes; previous small reports suggest that delays in ICU transfer adversely affect mortality and length of stay.12, 13, 24 Parkhe et al. compared 99 direct ICU admit patients with 23 who experienced early unplanned transfers; mortality at 30 days was significantly higher among patients with unplanned transfers.13 The current multifacility study included considerably more patients and confirmed an in‐hospital mortality gapalbeit a smaller onebetween patients with early transfers and those directly admitted to the ICU.
We focused on unplanned transfers during the earliest phase of hospitalization to identify patients who might benefit from improved recognition of, and intervention for, impending critical illness. We found that even patients requiring transfers within 8 hours of hospital admission were at an increased risk of death. Bapoje et al. recently reported that as many as 80% of early unplanned transfers were preventable and that most resulted from inappropriate admission triage.11 Together, these findings suggest that heightened attention to identifying such patients at admission or within the first day of hospitalizationwhen the rates of unplanned transfers peakis critical.
Several important limitations should be recognized in interpreting these results. First, this study was not designed to specifically identify the reasons for unplanned transfers, limiting our ability to characterize episodes in which timely care could have prevented excess mortality. Notably, while previous work suggests that many early unplanned transfers might be prevented with appropriate triage, it is likely that some excess deaths are not preventable even if every patient could be admitted to the ICU directly.
We were able to characterize patient outcomes by admitting diagnoses. Patients admitted for pneumonia and respiratory infection, gastrointestinal bleeding, COPD exacerbation, or seizures demonstrated excess mortality compared with matched ICU controls, while those with AMI, sepsis, and stroke did not. It is possible that differences in diagnosis‐specific excess mortality resulted from increasing adherence to well‐defined practice guidelines for specific high‐risk conditions.2527 For example, international awareness campaigns for the treatment of sepsis, AMI, and strokeSurviving Sepsis, Door‐to‐Balloon, and F.A.S.T.emphasize early interventions to minimize morbidity and mortality.
Second, the data utilized in this study were based on automated variables extracted from the electronic medical record. Mortality prediction models based on automated variables have demonstrated excellent performance among ICU and non‐ICU populations14, 18, 28; however, the inclusion of additional data (eg, vital signs or neurological status) would likely improve baseline risk adjustment.5, 10, 2931 Multiple studies have demonstrated that vital signs and clinician judgment can predict patients at an increased risk of deterioration.5, 10, 2931 Such data might also provide insight into residual factors that influenced clinicians' decisions to triage patients to an ICU versus non‐ICU admissiona focus area of our ongoing research efforts. Utilizing electronically available data, however, facilitated the identification of a cohort of patients far larger than that in prior studies. Where previous work has also been limited by substantial variability in baseline characteristics among study subjects,1, 2, 12, 13 our large sample produced a high percentage of multivariable case matches.
Third, we chose to match patients with a severity of illness index based on variables available at the time of hospital admission. While this mortality prediction model has demonstrated excellent performance in internal and external populations,14, 18 it is calibrated for general inpatient, rather than critically ill, populations. It remains possible that case matching with ICU‐specific severity of illness scores might alter matching characteristics, however, previous studies suggest that severity of illness, as measured by these scores, is comparable between direct ICU admits and early ICU transfers.13 Importantly, our matching procedure avoided the potential confounding known to exist with the use of prediction models based on discharge or intra‐hospitalization data.32, 33
Finally, while we were able to evaluate unplanned transfer timing in a multifacility sample, all patient care occurred within a large integrated healthcare delivery system. The overall observed mortality in our study was lower than that reported in prior studies which considered more limited patient cohorts.1, 2, 12, 13, 22 Thus, differences in patient case‐mix or ICU structure must be considered when applying our results to other healthcare delivery systems.
This hypothesis‐generating study, based on a large, multifacility sample of hospitalizations, suggests several areas of future investigation. Future work should detail specific aspects of care among patients with unplanned transfer, including: evaluating the structures and processes involved in triage decisions, measuring the effects on mortality through implementation of interventions (eg, rapid response teams or diagnosis‐specific treatment protocols), and defining the causes and risk factors for unplanned transfers by elapsed time.
In conclusion, the risk of an unplanned ICU transfera common event among hospitalized patientsis highest within 24 hours of hospitalization. Patients with early unplanned transfers have increased mortality and length of stay compared to those admitted directly to the ICU. Even patients transferred to the ICU within 8 hours of hospital admission are at an increased risk of death when compared with those admitted directly. Substantial variability in unplanned transfer outcomes exists based on admitting diagnoses. Future research should characterize unplanned transfers in greater detail with the goal of identifying patients that would benefit from improved triage and early ICU transfer.
- Admission source to the medical intensive care unit predicts hospital death independent of APACHE II score.JAMA.1990;264(18):2389–2394. , .
- Unplanned admission to intensive care after emergency hospitalisation: risk factors and development of a nomogram for individualising risk.Resuscitation.2009;80(2):224–230. , , , , , .
- Outcome of intensive care patients in a group of British intensive care units.Crit Care Med.1998;26(8):1337–1345. , .
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS).J Hosp Med.2010;6(2):74–80. , , , , , .
- Medical patients at high risk for catastrophic deterioration.Crit Care Med.1987;15(5):510–515. , .
- Introduction of the medical emergency team (MET) system: a cluster‐randomised controlled trial.Lancet.2005;365(9477):2091–2097. , , , et al.
- Effect of a rapid response team on hospital‐wide mortality and code rates outside the ICU in a children's hospital.JAMA.2007;298(19):2267–2274. , , , et al.
- Validity of unplanned admission to an intensive care unit as a measure of patient safety in surgical patients.Anesthesiology.2005;103(6):1121–1129. , , , , , .
- The 100,000 lives campaign: setting a goal and a deadline for improving health care quality.JAMA.2006;295(3):324–327. , , , .
- Duration of life‐threatening antecedents prior to intensive care admission.Intensive Care Med.2002;28(11):1629–1634. , , , et al.
- Unplanned transfers to a medical intensive care unit: causes and relationship to preventable errors in care.J Hosp Med.2011;6(2):68–72. , , , .
- Inpatient transfers to the intensive care unit: delays are associated with increased mortality and morbidity.J Gen Intern Med.2003;18(2):77–83. , , , , .
- Outcome of emergency department patients with delayed admission to an intensive care unit.Emerg Med (Fremantle).2002;14(1):50–57. , , , .
- Risk‐adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases.Med Care.2008;46(3):232–239. , , , , , .
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases.Am J Manag Care.2008;14(3):158–166. , , , et al.
- Linking automated databases for research in managed care settings.Ann Intern Med.1997;127(8 pt 2):719–724. .
- Anticoagulation therapy for stroke prevention in atrial fibrillation: how well do randomized trials translate into clinical practice?JAMA.2003;290(20):2685–2692. , , , et al.
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population.J Clin Epidemiol.2009;63(7):798–803. , , , .
- Refinements to the diagnostic cost group (DCG) model.Inquiry.1995;32(4):418–429. , .
- Excess length of stay, charges, and mortality attributable to medical injuries during hospitalization.JAMA.2003;290(14):1868–1874. , .
- Optimal matching in observational studies.J Am Stat Assoc.1989;84:1024–1032. .
- Admissions to intensive care units from emergency departments: a descriptive study.Emerg Med J.2005;22(6):423–428. , , , .
- Using administrative data to develop a nomogram for individualising risk of unplanned admission to intensive care.Resuscitation.2008;79(2):241–248. , , , .
- Unplanned intensive care unit transfers: a useful tool to improve quality of care [abstract]. In: Hospital Medicine 2010 abstract booklet. Society of Hospital Medicine 2010 Annual Meeting, April 9–11, 2010, Washington, DC;2010:10–11. , , , .
- Surviving Sepsis Campaign: international guidelines for management of severe sepsis and septic shock: 2008.Crit Care Med.2008;36(1):296–327. , , , et al.
- 2009 Focused Updates: ACC/AHA Guidelines for the Management of Patients With ST‐Elevation Myocardial Infarction (updating the 2004 Guideline and 2007 Focused Update) and ACC/AHA/SCAI Guidelines on Percutaneous Coronary Intervention (updating the 2005 Guideline and 2007 Focused Update): a report of the American College of Cardiology Foundation/American Heart Association Task Force on Practice Guidelines.Circulation.2009;120(22):2271–2306. , , , et al.
- Translating evidence into practice: a decade of efforts by the American Heart Association/American Stroke Association to reduce death and disability due to stroke: a presidential advisory from the American Heart Association/American Stroke Association.Stroke.2010;41(5):1051–1065. , , , et al.
- Veterans Affairs intensive care unit risk adjustment model: validation, updating, recalibration.Crit Care Med.2008;36(4):1031–1042. , , , et al.
- Recommended guidelines for monitoring, reporting, and conducting research on medical emergency team, outreach, and rapid response systems: an Utstein‐style scientific statement: a scientific statement from the International Liaison Committee on Resuscitation (American Heart Association, Australian Resuscitation Council, European Resuscitation Council, Heart and Stroke Foundation of Canada, InterAmerican Heart Foundation, Resuscitation Council of Southern Africa, and the New Zealand Resuscitation Council); the American Heart Association Emergency Cardiovascular Care Committee; the Council on Cardiopulmonary, Perioperative, and Critical Care; and the Interdisciplinary Working Group on Quality of Care and Outcomes Research.Circulation.2007;116(21):2481–2500. , , , et al.
- Realizing the potential of clinical judgment: a real‐time strategy for predicting outcomes and cost for medical inpatients.Am J Med.2000;109(3):189–195. , , , , , .
- Physiological values and procedures in the 24 h before ICU admission from the ward.Anaesthesia.1999;54(6):529–534. , , .
- Predicting who dies depends on how severity is measured: implications for evaluating patient outcomes.Ann Intern Med.1995;123(10):763–770. , , , , , .
- Enhancement of claims data to improve risk adjustment of hospital mortality.JAMA.2007;297(1):71–76. , , , et al.
- Admission source to the medical intensive care unit predicts hospital death independent of APACHE II score.JAMA.1990;264(18):2389–2394. , .
- Unplanned admission to intensive care after emergency hospitalisation: risk factors and development of a nomogram for individualising risk.Resuscitation.2009;80(2):224–230. , , , , , .
- Outcome of intensive care patients in a group of British intensive care units.Crit Care Med.1998;26(8):1337–1345. , .
- Intra‐hospital transfers to a higher level of care: contribution to total hospital and intensive care unit (ICU) mortality and length of stay (LOS).J Hosp Med.2010;6(2):74–80. , , , , , .
- Medical patients at high risk for catastrophic deterioration.Crit Care Med.1987;15(5):510–515. , .
- Introduction of the medical emergency team (MET) system: a cluster‐randomised controlled trial.Lancet.2005;365(9477):2091–2097. , , , et al.
- Effect of a rapid response team on hospital‐wide mortality and code rates outside the ICU in a children's hospital.JAMA.2007;298(19):2267–2274. , , , et al.
- Validity of unplanned admission to an intensive care unit as a measure of patient safety in surgical patients.Anesthesiology.2005;103(6):1121–1129. , , , , , .
- The 100,000 lives campaign: setting a goal and a deadline for improving health care quality.JAMA.2006;295(3):324–327. , , , .
- Duration of life‐threatening antecedents prior to intensive care admission.Intensive Care Med.2002;28(11):1629–1634. , , , et al.
- Unplanned transfers to a medical intensive care unit: causes and relationship to preventable errors in care.J Hosp Med.2011;6(2):68–72. , , , .
- Inpatient transfers to the intensive care unit: delays are associated with increased mortality and morbidity.J Gen Intern Med.2003;18(2):77–83. , , , , .
- Outcome of emergency department patients with delayed admission to an intensive care unit.Emerg Med (Fremantle).2002;14(1):50–57. , , , .
- Risk‐adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases.Med Care.2008;46(3):232–239. , , , , , .
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases.Am J Manag Care.2008;14(3):158–166. , , , et al.
- Linking automated databases for research in managed care settings.Ann Intern Med.1997;127(8 pt 2):719–724. .
- Anticoagulation therapy for stroke prevention in atrial fibrillation: how well do randomized trials translate into clinical practice?JAMA.2003;290(20):2685–2692. , , , et al.
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population.J Clin Epidemiol.2009;63(7):798–803. , , , .
- Refinements to the diagnostic cost group (DCG) model.Inquiry.1995;32(4):418–429. , .
- Excess length of stay, charges, and mortality attributable to medical injuries during hospitalization.JAMA.2003;290(14):1868–1874. , .
- Optimal matching in observational studies.J Am Stat Assoc.1989;84:1024–1032. .
- Admissions to intensive care units from emergency departments: a descriptive study.Emerg Med J.2005;22(6):423–428. , , , .
- Using administrative data to develop a nomogram for individualising risk of unplanned admission to intensive care.Resuscitation.2008;79(2):241–248. , , , .
- Unplanned intensive care unit transfers: a useful tool to improve quality of care [abstract]. In: Hospital Medicine 2010 abstract booklet. Society of Hospital Medicine 2010 Annual Meeting, April 9–11, 2010, Washington, DC;2010:10–11. , , , .
- Surviving Sepsis Campaign: international guidelines for management of severe sepsis and septic shock: 2008.Crit Care Med.2008;36(1):296–327. , , , et al.
- 2009 Focused Updates: ACC/AHA Guidelines for the Management of Patients With ST‐Elevation Myocardial Infarction (updating the 2004 Guideline and 2007 Focused Update) and ACC/AHA/SCAI Guidelines on Percutaneous Coronary Intervention (updating the 2005 Guideline and 2007 Focused Update): a report of the American College of Cardiology Foundation/American Heart Association Task Force on Practice Guidelines.Circulation.2009;120(22):2271–2306. , , , et al.
- Translating evidence into practice: a decade of efforts by the American Heart Association/American Stroke Association to reduce death and disability due to stroke: a presidential advisory from the American Heart Association/American Stroke Association.Stroke.2010;41(5):1051–1065. , , , et al.
- Veterans Affairs intensive care unit risk adjustment model: validation, updating, recalibration.Crit Care Med.2008;36(4):1031–1042. , , , et al.
- Recommended guidelines for monitoring, reporting, and conducting research on medical emergency team, outreach, and rapid response systems: an Utstein‐style scientific statement: a scientific statement from the International Liaison Committee on Resuscitation (American Heart Association, Australian Resuscitation Council, European Resuscitation Council, Heart and Stroke Foundation of Canada, InterAmerican Heart Foundation, Resuscitation Council of Southern Africa, and the New Zealand Resuscitation Council); the American Heart Association Emergency Cardiovascular Care Committee; the Council on Cardiopulmonary, Perioperative, and Critical Care; and the Interdisciplinary Working Group on Quality of Care and Outcomes Research.Circulation.2007;116(21):2481–2500. , , , et al.
- Realizing the potential of clinical judgment: a real‐time strategy for predicting outcomes and cost for medical inpatients.Am J Med.2000;109(3):189–195. , , , , , .
- Physiological values and procedures in the 24 h before ICU admission from the ward.Anaesthesia.1999;54(6):529–534. , , .
- Predicting who dies depends on how severity is measured: implications for evaluating patient outcomes.Ann Intern Med.1995;123(10):763–770. , , , , , .
- Enhancement of claims data to improve risk adjustment of hospital mortality.JAMA.2007;297(1):71–76. , , , et al.
Copyright © 2011 Society of Hospital Medicine
Continuing Medical Education Program in
If you wish to receive credit for this activity, which begins on the next page, please refer to the website:
Accreditation and Designation Statement
Blackwell Futura Media Services designates this educational activity for a 1 AMA PRA Category 1 Credit. Physicians should only claim credit commensurate with the extent of their participation in the activity.
Blackwell Futura Media Services is accredited by the Accreditation Council for Continuing Medical Education to provide continuing medical education for physicians.
Educational Objectives
Upon completion of this educational activity, participants will be better able to employ automated bed history data to examine outcomes of intra‐hospital transfers using all hospital admissions as the denominator.
Continuous participation in the Journal of Hospital Medicine CME program will enable learners to be better able to:
-
Interpret clinical guidelines and their applications for higher quality and more efficient care for all hospitalized patients.
-
Describe the standard of care for common illnesses and conditions treated in the hospital; such as pneumonia, COPD exacerbation, acute coronary syndrome, HF exacerbation, glycemic control, venous thromboembolic disease, stroke, etc.
-
Discuss evidence‐based recommendations involving transitions of care, including the hospital discharge process.
-
Gain insights into the roles of hospitalists as medical educators, researchers, medical ethicists, palliative care providers, and hospital‐based geriatricians.
-
Incorporate best practices for hospitalist administration, including quality improvement, patient safety, practice management, leadership, and demonstrating hospitalist value.
-
Identify evidence‐based best practices and trends for both adult and pediatric hospital medicine.
Instructions on Receiving Credit
For information on applicability and acceptance of continuing medical education credit for this activity, please consult your professional licensing board.
This activity is designed to be completed within the time designated on the title page; physicians should claim only those credits that reflect the time actually spent in the activity. To successfully earn credit, participants must complete the activity during the valid credit period that is noted on the title page.
Follow these steps to earn credit:
-
Log on to
www.blackwellpublishing.com/cme . -
Read the target audience, learning objectives, and author disclosures.
-
Read the article in print or online format.
-
Reflect on the article.
-
Access the CME Exam, and choose the best answer to each question.
-
Complete the required evaluation component of the activity.
If you wish to receive credit for this activity, which begins on the next page, please refer to the website:
Accreditation and Designation Statement
Blackwell Futura Media Services designates this educational activity for a 1 AMA PRA Category 1 Credit. Physicians should only claim credit commensurate with the extent of their participation in the activity.
Blackwell Futura Media Services is accredited by the Accreditation Council for Continuing Medical Education to provide continuing medical education for physicians.
Educational Objectives
Upon completion of this educational activity, participants will be better able to employ automated bed history data to examine outcomes of intra‐hospital transfers using all hospital admissions as the denominator.
Continuous participation in the Journal of Hospital Medicine CME program will enable learners to be better able to:
-
Interpret clinical guidelines and their applications for higher quality and more efficient care for all hospitalized patients.
-
Describe the standard of care for common illnesses and conditions treated in the hospital; such as pneumonia, COPD exacerbation, acute coronary syndrome, HF exacerbation, glycemic control, venous thromboembolic disease, stroke, etc.
-
Discuss evidence‐based recommendations involving transitions of care, including the hospital discharge process.
-
Gain insights into the roles of hospitalists as medical educators, researchers, medical ethicists, palliative care providers, and hospital‐based geriatricians.
-
Incorporate best practices for hospitalist administration, including quality improvement, patient safety, practice management, leadership, and demonstrating hospitalist value.
-
Identify evidence‐based best practices and trends for both adult and pediatric hospital medicine.
Instructions on Receiving Credit
For information on applicability and acceptance of continuing medical education credit for this activity, please consult your professional licensing board.
This activity is designed to be completed within the time designated on the title page; physicians should claim only those credits that reflect the time actually spent in the activity. To successfully earn credit, participants must complete the activity during the valid credit period that is noted on the title page.
Follow these steps to earn credit:
-
Log on to
www.blackwellpublishing.com/cme . -
Read the target audience, learning objectives, and author disclosures.
-
Read the article in print or online format.
-
Reflect on the article.
-
Access the CME Exam, and choose the best answer to each question.
-
Complete the required evaluation component of the activity.
If you wish to receive credit for this activity, which begins on the next page, please refer to the website:
Accreditation and Designation Statement
Blackwell Futura Media Services designates this educational activity for a 1 AMA PRA Category 1 Credit. Physicians should only claim credit commensurate with the extent of their participation in the activity.
Blackwell Futura Media Services is accredited by the Accreditation Council for Continuing Medical Education to provide continuing medical education for physicians.
Educational Objectives
Upon completion of this educational activity, participants will be better able to employ automated bed history data to examine outcomes of intra‐hospital transfers using all hospital admissions as the denominator.
Continuous participation in the Journal of Hospital Medicine CME program will enable learners to be better able to:
-
Interpret clinical guidelines and their applications for higher quality and more efficient care for all hospitalized patients.
-
Describe the standard of care for common illnesses and conditions treated in the hospital; such as pneumonia, COPD exacerbation, acute coronary syndrome, HF exacerbation, glycemic control, venous thromboembolic disease, stroke, etc.
-
Discuss evidence‐based recommendations involving transitions of care, including the hospital discharge process.
-
Gain insights into the roles of hospitalists as medical educators, researchers, medical ethicists, palliative care providers, and hospital‐based geriatricians.
-
Incorporate best practices for hospitalist administration, including quality improvement, patient safety, practice management, leadership, and demonstrating hospitalist value.
-
Identify evidence‐based best practices and trends for both adult and pediatric hospital medicine.
Instructions on Receiving Credit
For information on applicability and acceptance of continuing medical education credit for this activity, please consult your professional licensing board.
This activity is designed to be completed within the time designated on the title page; physicians should claim only those credits that reflect the time actually spent in the activity. To successfully earn credit, participants must complete the activity during the valid credit period that is noted on the title page.
Follow these steps to earn credit:
-
Log on to
www.blackwellpublishing.com/cme . -
Read the target audience, learning objectives, and author disclosures.
-
Read the article in print or online format.
-
Reflect on the article.
-
Access the CME Exam, and choose the best answer to each question.
-
Complete the required evaluation component of the activity.
Intra‐Hospital Transfer to a Higher Level of Care
Considerable research and public attention is being paid to the quantification, risk adjustment, and reporting of inpatient mortality.15 Inpatient mortality is reported as aggregate mortality (for all hospitalized patients or those with a specific diagnosis3, 6) or intensive care unit (ICU) mortality.7, 8 While reporting aggregate hospital or aggregate ICU mortality rates is useful, it is also important to develop reporting strategies that go beyond simply using data elements found in administrative databases (eg, diagnosis and procedure codes) to quantify practice variation. Ideally, such strategies would permit delineating processes of careparticularly those potentially under the control of hospitalists, not only intensiviststo identify improvement opportunities. One such process, which can be tracked using the bed history component of a patient's electronic medical record, is the transfer of patients between different units within the same hospital.
Several studies have documented that risk of ICU death is highest among patients transferred from general medical‐surgical wards, intermediate among direct admissions from the emergency department, and lowest among surgical admissions.911 Opportunities to reduce subsequent ICU mortality have been studied among ward patients who develop sepsis and are then transferred to the ICU,12 among patients who experience cardiac arrest,13, 14 as well as among patients with any physiological deterioration (eg, through the use of rapid response teams).1517 Most of these studies have been single‐center studies and/or studies reporting only an ICU denominator. While useful in some respects, such studies are less helpful to hospitalists, who would benefit from better understanding of the types of patients transferred and the total impact that transfers to a higher level of care make on general medical‐surgical wards. In addition, entities such as the Institute for Healthcare Improvement recommend the manual review of records of patients who were transferred from the ward to the ICU18 to identify performance improvement opportunities. While laudable, such approaches do not lend themselves to automated reporting strategies.
We recently described a new risk adjustment methodology for inpatient mortality based entirely on automated data preceding hospital admission and not restricted to ICU patients. This methodology, which has been externally validated in Ottawa, Canada, after development in the Kaiser Permanente Medical Care Program (KPMCP), permits quantification of a patient's pre‐existing comorbidity burden, physiologic derangement at the time of admission, and overall inpatient mortality risk.19, 20 The primary purpose of this study was to combine this methodology with bed history analysis to quantify the in‐hospital mortality and length of stay (LOS) of patients who experienced intra‐hospital transfers in a large, multihospital system. As a secondary goal, we also wanted to assess the degree to which these transfers could be predicted based on information available prior to a patient's admission.
ABBREVIATIONS AND TERMS USED IN TEXT
COPS: COmorbidity Point Score. Point score based on a patient's health care utilization diagnoses (during the year preceding admission to the hospital. Analogous to POA (present on admission) coding. Scores can range from 0 to a theoretical maximum of 701 but scores >200 are rare. With respect to a patient's pre‐existing comorbidity burden, the unadjusted relationship of COPS and inpatient mortality is as follows: a COPS <50 is associated with a mortality risk of <1%, <100 with a mortality risk of <5%, 100 to 145 with a mortality risk of 5% to 10%, and >145 with a mortality risk of 10% or more.
ICU: Intensive Care Unit. In this study, all ICUs have a minimum registered nurse to patient ratio of 1:2.
LAPS: Laboratory Acute Physiology Score. Point score based on 14 laboratory test results obtained in the 72 hours preceding hospitalization. With respect to a patient's physiologic derangement, the unadjusted relationship of LAPS and inpatient mortality is as follows: a LAPS <7 is associated with a mortality risk of <1%, <7 to 30 with a mortality risk of <5%, 30 to 60 with a mortality risk of 5% to 9%, and >60 with a mortality risk of 10% or more.
LOS: Exact hospital Length Of Stay. LOS is calculated from admission until first discharge home (i.e., it may span more than one hospital stay if a patient experienced inter‐hospital transport).
Predicted (expected) mortality risk: the % risk of death for a given patient based on his/her age, sex, admission diagnosis, COPS, and LAPS.
OEMR: Observed to Expected Mortality Ratio. For a given patient subset, the ratio of the actual mortality experienced by the subset to the expected (predicted) mortality for the subset. Predicted mortality is based on patients' age, sex, admission diagnosis, COPS, and LAPS.
OMELOS: Observed Minus Expected LOS. For a given patient subset, the difference between the actual number of hospital days experienced by the subset and the expected (predicted) number of hospital days for the subset. Predicted LOS is based on patients' age, sex, admission diagnosis, COPS, and LAPS.
TCU: Transitional Care Unit (also called intermediate care unit or stepdown unit). In this study, TCUs have variable nurse to patient ratios ranging from 1:2.5 to 1:3 and did not provide assisted ventilation, continuous pressor infusions, or invasive monitoring.
Materials and Methods
This project was approved by the Northern California KPMCP Institutional Review Board for the Protection of Human Subjects.
The Northern California KPMCP serves a total population of approximately 3.3 million members. Under a mutual exclusivity arrangement, physicians of The Permanente Medical Group, Inc., care for Kaiser Foundation Health Plan, Inc. members at facilities owned by Kaiser Foundation Hospitals, Inc. All Northern California KPMCP hospitals and clinics employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere. Databases maintained by the KPMCP capture admission and discharge times, admission and discharge diagnoses and procedures (assigned by professional coders), bed histories, inter‐hospital transfers, as well as the results of all inpatient and outpatient laboratory tests. The use of these databases for research has been described in multiple reports.2124
Our setting consisted of all 19 hospitals owned and operated by the KPMCP, whose characteristics are summarized in the Supporting Information Appendix available to interested readers. These include the 17 described in our previous report19 as well as 2 new hospitals (Antioch and Manteca) which are similar in size and type of population served. Our study population consisted of all patients admitted to these 19 hospitals who met these criteria: 1) hospitalization began from November 1st, 2006 through January 31st, 2008; 2) initial hospitalization occurred at a Northern California KPMCP hospital (ie, for inter‐hospital transfers, the first hospital stay occurred within the KPMCP); 3) age 15 years; and 4) hospitalization was not for childbirth.
We defined a linked hospitalization as the time period that began with a patient's admission to the hospital and ended with the patient's discharge (home, to a nursing home, or death). Linked hospitalizations can thus involve more than 1 hospital stay and could include a patient transfer from one hospital to another prior to definitive discharge. For linked hospitalizations, mortality was attributed to the admitting KPMCP hospital (ie, if a patient was admitted to hospital A, transferred to B, and died at hospital B, mortality was attributed to hospital A). We defined total LOS as the exact time in hours from when a patient was first admitted to the hospital until death or final discharge home or to a nursing home, while total ICU or transitional care unit (TCU, referred to as stepdown unit in some hospitals) LOS was calculated for all individual ICU or TCU stays during the hospital stay.
Intra‐Hospital Transfers
We grouped all possible hospital units into four types: general medical‐surgical ward (henceforth, ward); operating room (OR)/post‐anesthesia recovery (PAR); TCU; and ICU. In 2003, the KPMCP implemented a mandatory minimum staffing ratio of one registered nurse for every four patients in all its hospital units; in addition, staffing levels for designated ICUs adhered to the previously mandated minimum of one nurse for every 2 patients. So long as they adhere to these minimum ratios, individual hospitals have considerable autonomy with respect to how they staff or designate individual hospital units. Registered nurse‐to‐patient ratios during the time of this study were as follows: ward patients, 1:3.5 to 1:4; TCU patients, 1:2.5 to 1:3; and ICU patients, 1:1 to 1:2. Staffing ratios for the OR and PAR are more variable, depending on the surgical procedures involved. Current KPMCP databases do not permit accurate quantification of physician staffing. All 19 study hospitals had designated ICUs, 6 were teaching hospitals, and 11 had designated TCUs. None of the study hospitals had closed ICUs (units where only intensivists admit patients) and none had continuous coverage of the ICU by intensivists. While we were not able to employ electronic data to determine who made the decision to transfer, we did find considerable variation with respect to how intensivists covered the ICUs and how they interfaced with hospitalists. Staffing levels for specialized coronary care units and non‐ICU monitored beds were not standardized. All study hospitals had rapid response teams as well as code blue teams during the time period covered by this report. Respiratory care practitioners were available to patients in all hospital units, but considerable variation existed with respect to other services available (eg, cardiac catheterization units, provision of noninvasive positive pressure ventilation outside the ICU, etc.).
This report focuses on intra‐hospital transfers to the ICU and TCU, with special emphasis on nonsurgical transfers (due to space limitations, we are not reporting on the outcomes of patients whose first hospital unit was the OR; additional details on these patients are provided in the Supporting Information Appendix). For the purposes of this report, we defined the following admission types: direct admits (patients admitted to the ICU or TCU whose first hospital unit on admission was the ICU or TCU); and nonsurgical transfers to a higher level of care. These latter transfers could be of 3 types: ward to ICU, ward to TCU, and TCU to ICU. We also quantified the effect of inter‐hospital transfers.
Independent Variables
In addition to patients' age and sex, we employed the following independent variables to predict transfer to a higher level of care. These variables are part of the risk adjustment model described in greater detail in our previous report19 and were available electronically for all patients in the cohort. We grouped admission diagnoses into 44 broad diagnostic categories (Primary Conditions), and admission types into 4 groups (emergency medical, emergency surgical, elective medical, and elective surgical). We quantified patients' degree of physiologic derangement using a Laboratory‐based Acute Physiology Score (LAPS) using laboratory test results prior to hospitalization. We quantified patients' comorbid illness burden using a Comorbidity Point Score (COPS) based on patients' pre‐existing diagnoses over the 12‐month period preceding hospitalization. Lastly, we assigned each patient a predicted mortality risk (%) and LOS based on the above predictors,19 permitting calculation of observed to expected mortality ratios (OEMRs) and observed minus expected LOS (OMELOS).
Statistical Methods
All analyses were performed in SAS.25 We calculated standard descriptive statistics (medians, means, standard deviations) and compared different patient groupings using t and chi‐square tests. We employed a similar approach to that reported by Render et al.7 to calculate OEMR and OMELOS.
To determine the degree to which transfers to a higher level of care from the ward or TCU would be predictable using information available at the time of admission, we performed 4 sets of logistic regression analyses using the above‐mentioned predictors in which the outcome variables were as follows: 1) transfer occurring in the first 48 hours after admission (time frame by which point approximately half of the transferred patients experienced a transfer) among ward or TCU patients and 2) transfer occurring after 48 hours among ward or TCU patients. We evaluated the discrimination and calibration of these models using the same methods described in our original report (measuring the area under the receiver operator characteristic curve, or c statistic, and visually examining observed and expected mortality rates among predicted risk bands as well as risk deciles) as well as additional statistical tests recommended by Cook.19, 26
Results
During the study period, a total of 249,129 individual hospital stays involving 170,151 patients occurred at these 19 hospitals. After concatenation of inter‐hospital transfers, we were left with 237,208 linked hospitalizations. We excluded 26,738 linked hospitalizations that began at a non‐KPMCP hospital (ie, they were transported in), leaving a total of 210,470 linked hospitalizations involving 150,495 patients. The overall linked hospitalization mortality rate was 3.30%.
Table 1 summarizes cohort characteristics based on initial hospital location. On admission, ICU patients had the highest degree of physiologic derangement as well as the highest predicted mortality. Considerable inter‐hospital variation was present in both predictors and outcomes; details on these variations are provided in the Supporting Information Appendix.
Ward | TCU | ICU | All* | |
---|---|---|---|---|
| ||||
n | 121,237 | 20,556 | 16,001 | 210,470 |
Admitted via emergency department, n (%) | 99,909 (82.4) | 18,612 (90.5) | 13,847 (86.5) | 139,036 (66.1) |
% range across hospitals | 55.0‐94.2 | 64.7‐97.6 | 49.5‐97.4 | 53.6‐76.9 |
Male, n (%) | 53,744 (44.3) | 10,362 (50.4) | 8,378 (52.4) | 94,451 (44.9) |
Age in years (mean SD) | 64.5 19.2 | 69.0 15.6 | 63.7 17.8 | 63.2 18.6 |
LAPS (mean SD) | 19.2 18.0 | 23.3 19.5 | 31.7 25.7 | 16.7 19.0 |
COPS (mean SD) | 90.4 64.0 | 99.2 65.9 | 94.5 67.5 | 84.7 61.8 |
% predicted mortality (mean SD) | 4.0 7.1 | 4.6 7.3 | 8.7 12.8 | 3.6 7.3 |
Observed in‐hospital deaths (n, %) | 3,793 (3.1) | 907 (4.4) | 1,995 (12.5) | 6,952 (3.3) |
Observed to expected mortality ratio | 0.79 (0.77‐0.82) | 0.95 (0.89‐1.02) | 1.43 (1.36‐1.49) | 0.92 (0.89‐0.94) |
Total hospital LOS, days (mean SD) | 4.6 7.5 | 5.3 10.0 | 7.8 14.0 | 4.6 8.1 |
Table 2 summarizes data from 3 groups of patients: patients initially admitted to the ward, or TCU, who did not experience a transfer to a higher level of care and patients admitted to these 2 units who did experience such a transfer. Patients who experienced a transfer constituted 5.3% (6,484/121,237) of ward patients and 6.7% (1,384/20,556) of TCU patients. Transferred patients tended to be older, have more acute physiologic derangement (higher LAPS), a greater pre‐existing illness burden (higher COPS), and a higher predicted mortality risk. Among ward patients, those with the following admission diagnoses were most likely to experience a transfer to a higher level of care: gastrointestinal bleeding (10.8% of all transfers), pneumonia (8.7%), and other infections (8.2%). The diagnoses most likely to be associated with death following transfer were cancer (death rate among transferred patients, 48%), renal disease (death rate, 36%), and liver disease (33%). Similar distributions were observed for TCU patients.
Patients Initially Admitted to Ward, Remained There | Patients Initially Admitted to TCU, Remained There | Patients Transferred to Higher Level of Care | All | |
---|---|---|---|---|
| ||||
n | 114,753 | 19,172 | 7,868 | 141,793 |
Male, n (%) | 50,586 (44.1) | 9,626 (50.2) | 3,894 (49.5) | 64,106 (45.2) |
Age (mean SD) | 64.3 19.4 | 69.0 15.7 | 68.1 16.1 | 65.2 18.8 |
LAPS (mean SD) | 18.9 17.8 | 22.7 19.1 | 26.7 21.0 | 19.8 18.3 |
COPS (mean SD) | 89.4 63.7 | 98.3 65.5 | 107.9 67.6 | 91.7 64.4 |
% predicted mortality risk (mean SD) | 3.8 7.0 | 4.4 7.0 | 6.5 8.8 | 4.1 7.1 |
Admission diagnosis of pneumonia, n (%) | 5,624 (4.9) | 865 (4.5) | 684 (8.7) | 7,173 (5.1) |
Admission diagnosis of sepsis, n (%) | 1,181 (1.0) | 227 (1.2) | 168 (2.1) | 1,576 (1.1) |
Admission diagnosis of GI bleed, n (%) | 13,615 (11.9) | 1,448 (7.6) | 851 (10.8) | 15,914 (11.2) |
Admission diagnosis of cancer, n (%) | 2,406 (2.1) | 80 (0.4) | 186 (2.4) | 2,672 (1.9) |
Table 3 compares outcomes among ward and TCU patients who did and did not experience a transfer to a higher level of care. The table shows that transferred patients were almost 3 times as likely to die, even after controlling for severity of illness, and that their hospital LOS was 9 days higher than expected. This increased risk was seen in all hospitals and among all transfer types (ward to ICU, ward to TCU, and TCU to ICU).
Patients Initially Admitted to Ward, Remained There | Patients Initially Admitted to TCU, Remained There | Patients Transferred to Higher Level of Care | |
---|---|---|---|
| |||
n | 114,753 | 19,172 | 7,868 |
Admitted to ICU, n (%) | 0 (0.0) | 0 (0.0) | 5,245 (66.7) |
Ventilated, n (%) | 0 (0.0) | 0 (0.0) | 1,346 (17.1) |
Died in the hospital, n (%) | 2,619 (2.3) | 572 (3.0) | 1,509 (19.2) |
Length of stay, in days, at time of death (mean SD) | 7.0 11.9 | 8.3 12.4 | 16.2 23.7 |
Observed to expected mortality ratio (95% CI) | 0.60 (0.57‐0.62) | 0.68 (0.63‐0.74) | 2.93 (2.79‐3.09) |
Total hospital length of stay, days (mean SD) | 4.0 5.7 | 4.4 6.9 | 14.3 21.3 |
Observed minus expected length of stay (95% CI) | 0.4 (0.3‐0.4) | 0.8 (0.7‐0.9) | 9.1 (8.6‐9.5) |
Length of stay, in hours, at time of transfer (mean SD) | 80.8 167.2 |
Table 3 also shows that, among decedent patients, those who never left the ward or TCU died much sooner than those who died following transfer. Among direct admits to the ICU, the median LOS at time of death was 3.9 days, with a mean of 9.4 standard deviation of 19.9 days, while the corresponding times for TCU direct admits were a median and mean LOS of 6.5 and 11.7 19.5 days.
Table 4 summarizes outcomes among different patient subgroups that did and did not experience a transfer to a higher level of care. Based on location, patients who experienced a transfer from the TCU to the ICU had the highest crude death rate, but patients transferred from the ward to the ICU had the highest OEMR. On the other hand, if one divides patients by the degree of physiologic derangement, patients with low LAPS who experienced a transfer had the highest OEMR. With respect to LOS, patients transferred from the TCU to the ICU had the highest OMELOS (13.4 extra days).
n (%)* | Death Rate (%) | OEMR | LOS (mean SD) | OMELOS | |
---|---|---|---|---|---|
| |||||
Never admitted to TCU or ICU | 157,632 (74.9) | 1.6 | 0.55 (0.53‐0.57) | 3.6 4.6 | 0.04 (0.02‐0.07) |
Direct admit to TCU | 18,464 (8.8) | 2.9 | 0.66 (0.61‐0.72) | 4.2 5.8 | 0.60 (0.52‐0.68) |
Direct admit to ICU | 14,655 (7.0) | 11.9 | 1.38 (1.32‐1.45) | 6.4 9.4 | 2.28 (2.14‐2.43) |
Transferred from ward to ICU | 5,145 (2.4) | 21.5 | 3.23 (3.04‐3.42) | 15.7 21.6 | 10.33 (9.70‐10.96) |
Transferred from ward to TCU | 3,144 (1.5) | 11.9 | 1.99 (1.79‐2.20) | 13.6 23.2 | 8.02 (7.23‐8.82) |
Transferred from TCU to ICU | 1,107 (0.5) | 25.7 | 2.94 (2.61‐3.31) | 18.0 28.2 | 13.35 (11.49‐15.21) |
Admitted to ward, COPS 80, no transfer to ICU or TCU | 55,405 (26.3) | 3.4 | 0.59 (0.56‐0.62) | 4.5 5.9 | 0.29 (0.24‐0.34) |
Admitted to ward, COPS 80, did experience transfer to ICU or TCU | 4,851 (2.3) | 19.3 | 2.72 (2.55‐2.90) | 14.2 20.0 | 8.14 (7.56‐8.71) |
Admitted to ward, COPS <80, no transfer to ICU or TCU | 57,421 (27.3) | 1.1 | 0.55 (0.51‐0.59) | 3.4 4.2 | 0.23 (0.19‐0.26) |
Admitted to ward, COPS <80, did experience transfer to ICU or TCU | 3,560 (1.7) | 9.8 | 2.93 (2.63‐3.26) | 12.0 19.0 | 7.52 (6.89‐8.15) |
Admitted to ward, LAPS 20, no transfer to ICU or TCU | 46,492 (22.1) | 4.2 | 0.59 (0.56‐0.61) | 4.6 5.4 | 0.16 (0.12‐0.21) |
Admitted to ward, LAPS 20, did experience transfer to ICU or TCU | 4,070 (1.9) | 21.4 | 2.37 (2.22‐2.54) | 14.8 21.0 | 8.76 (8.06‐9.47) |
Admitted to ward, LAPS <20, no transfer to ICU or TCU | 66,334 (31.5) | 0.9 | 0.55 (0.51‐0.60) | 3.5 4.9 | 0.32 (0.28‐0.36) |
Admitted to ward, LAPS <20, did experience transfer to ICU or TCU | 4,341 (2.1) | 9.5 | 4.31 (3.90‐4.74) | 11.8 18.1 | 7.12 (6.61‐7.64) |
Transfers to a higher level of care at a different hospital, which in the KPMCP are usually planned, experienced lower mortality than transfers within the same hospital. For ward to TCU transfers, intra‐hospital transfers had a mortality of 12.1% while inter‐hospital transfers had a mortality of 5.7%. Corresponding rates for ward to ICU transfers were 21.7% and 11.2%, and for TCU to ICU transfers the rates were 25.9% and 12.5%, respectively.
Among patients initially admitted to the ward, a model to predict the occurrence of a transfer to a higher level of care (within 48 hours after admission) that included age, sex, admission type, primary condition, LAPS, COPS, and interaction terms had poor discrimination, with an area under the receiver operator characteristic (c statistic) of only 0.64. The c statistic for a model to predict transfer after 48 hours was 0.66. The corresponding models for TCU admits had c statistics of 0.67 and 0.68. All four models had poor calibration.
Discussion
Using automated bed history data permits characterizing a patient population with disproportionate mortality and LOS: intra‐hospital transfers to special care units (ICUs or TCUs). Indeed, the largest subset of these patients (those initially admitted to the ward or TCU) constituted only 3.7% of all admissions, but accounted for 24.2% of all ICU admissions, 21.7% of all hospital deaths, and 13.2% of all hospital days. These patients also had very elevated OEMRs and OMELOS. Models based on age, sex, preadmission laboratory test results, and comorbidities did not predict the occurrence of these transfers.
We performed multivariate analyses to explore the degree to which electronically assigned preadmission severity scores could predict these transfers. These analyses found that, compared to our ability to predict inpatient or 30‐day mortality at the time of admission, which is excellent, our ability to predict the occurrence of transfer after admission is much more limited. These results highlight the limitations of severity scores that rely on automated data, which may not have adequate discrimination when it comes to determining the risk of an adverse outcome within a narrow time frame. For example, among the 121,237 patients initially admitted to the ward who did not experience an intra‐hospital transfer, the mean LAPS was 18.9, while the mean LAPS among the 6,484 ward patients who did experience a transfer was 25.5. Differences between the mean and median LAPS, COPS, and predicted mortality risk among transferred and non‐transferred patients were significant (P < 0.0001 for all comparisons). However, examination of the distribution of LAPS, COPS, and predicted mortality risk between these two groups of patients showed considerable overlap.
Our methodology resembles Silber et al.'s27, 28 concept of failure to rescue in that it focuses on events occurring after hospitalization. Silber et al. argue that a hospital's quality can be measured by quantifying the degree to which patients who experience new problems are successfully rescued. Furthermore, quantification of those situations where rescue attempts are unsuccessful is felt to be superior to simply comparing raw or adjusted mortality rates because these are primarily determined by underlying case mix. The primary difference between Silber et al.'s approach and ours is at the level of detailthey specified a specific set of complications, whereas our measure is more generic and would include patients with many of the complications specified by Silber et al.27, 28
Most of the patients transferred to a higher level of care in our cohort survived (ie, were rescued), indicating that intensive care is beneficial. However, the fact that these patients had elevated OEMRs and OMELOS indicates that the real challenge facing hospitalists involves the timing of provision of a beneficial intervention. In theory, improved timing could result from earlier detection of problems, which is the underlying rationale for employing rapid response teams. However, the fact that our electronic tools (LAPS, COPS) cannot predict patient deteriorations within a narrow time frame suggests that early detection will remain a major challenge. Manually assigned vital signs scores designed for this purpose do not have good discrimination either.29, 30 This raises the possibility that, though patient groups may differ in terms of overall illness severity and mortality risk, differences at the individual patient level may be too subtle for clinicians to detect. Future research may thus need to focus on scores that combine laboratory data, vital signs, trends in data,31, 32 and newer proteomic markers (eg, procalcitonin).33 We also found that most transfers occurred early (within <72 hours), raising the possibility that at least some of these transfers may involve issues around triage rather than sudden deterioration.
Our study has important limitations. Due to resource constraints and limited data availability, we could not characterize the patients as well as might be desirable; in particular, we could not make full determinations of the actual reasons for patients' transfer for all patients. Broadly speaking, transfer to a higher level of care could be due to inappropriate triage, appropriate (preventive) transfer (which could include transfer to a more richly staffed unit for a specific procedure), relentless progression of disease despite maximal therapy, the occurrence of management errors, patient and family uncertainty about goals of care or inadequate understanding of treatment options and prognoses, or a combination of these factors. We could not make these distinctions with currently available electronic data. This is also true of postsurgical patients, in whom it is difficult to determine which transfers to intensive care might be planned (eg, in the case of surgical procedures where ICU care is anticipated) as opposed to the occurrence of a deterioration during or following surgery. Another major limitation of this study is our inability to identify code or no code status electronically. The elapsed LOS at time of death among patients who experienced a transfer to a higher level of care (as compared to patients who died in the ward without ever experiencing intra‐hospital transfer) suggests, but does not prove, that prolonged efforts were being made to keep them alive. We were also limited in terms of having access to other process data (eg, physician staffing levels, provision and timing of palliative care). Having ICU severity of illness scores would have permitted us to compare our cohort to those of other recent studies showing elevated mortality rates among transfer patients,911 but we have not yet developed that capability.
Consideration of our study findings suggests a possible research agenda that could be implemented by hospitalist researchers. This agenda should emphasize three areas: detection, intervention, and reflection.
With respect to detection, attention needs to be paid to better tools for quantifying patient risk at the time a decision to admit to the ward is made. It is likely that such tools will need to combine the attributes of our severity score (LAPS) with those of the manually assigned scores.30, 34 In some cases, use of these tools could lead a physician to change the locus of admission from the ward to the TCU or ICU, which could improve outcomes by ensuring more timely provision of intensive care. Since problems with initial triage could be due to factors other than the failure to suspect or anticipate impending instability, future research should also include a cognitive component (eg, quantifying what proportion of subsequent patient deteriorations could be ascribed to missed diagnoses35). Additional work also needs to be done on developing mathematical models that can inform electronic monitoring of ward (not just ICU) patients.
Research on interventions that hospitalists can use to prevent the need for intensive care or to improve the rescue rate should take two routes. The first is a disease‐specific route, which builds on the fact that a relatively small set of conditions (pneumonia, sepsis, gastrointestinal bleeding) account for most transfers to a higher level of care. Condition‐specific protocols, checklists, and bundles36 tailored to a ward environment (as opposed to the ICU or to the entire hospital) might prevent deteriorations in these patients, as has been reported for sepsis.37 The second route is to improve the overall capabilities of rapid response and code blue teams. Such research would need to include a more careful assessment of what commonalities exist among patients who were and were not successfully rescued by these teams. This approach would probably yield more insights than the current literature, which focuses on whether rapid response teams are a good thing or not.
Finally, research also needs to be performed on how hospitalists reflect on adverse outcomes among ward patients. Greater emphasis needs to be placed on moving beyond trigger tool approaches that rely on manual chart review. In an era of expanding use of electronic medical record systems, more work needs to be done on how to harness these to provide hospitalists with better quantitative and risk‐adjusted information. This information should not be limited to simply reporting rates of transfers and deaths. Rather, finer distinctions must be provided with respect of the type of patients (ie, more diagnostic detail), the clinical status of patients (ie, more physiologic detail), as well as the effects of including or excluding patients in whom therapeutic options may be limited (ie, do not resuscitate and comfort care patients) on reported rates. Ideally, researchers should develop better process and outcomes measures that could be tested in collaborative networks that include multiple nonacademic general medical‐surgical wards.
Acknowledgements
The authors thank Drs. Paul Feigenbaum, Alan Whippy, Joseph V. Selby, and Philip Madvig for reviewing the manuscript and Ms. Jennifer Calhoun for formatting the manuscript.
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Considerable research and public attention is being paid to the quantification, risk adjustment, and reporting of inpatient mortality.15 Inpatient mortality is reported as aggregate mortality (for all hospitalized patients or those with a specific diagnosis3, 6) or intensive care unit (ICU) mortality.7, 8 While reporting aggregate hospital or aggregate ICU mortality rates is useful, it is also important to develop reporting strategies that go beyond simply using data elements found in administrative databases (eg, diagnosis and procedure codes) to quantify practice variation. Ideally, such strategies would permit delineating processes of careparticularly those potentially under the control of hospitalists, not only intensiviststo identify improvement opportunities. One such process, which can be tracked using the bed history component of a patient's electronic medical record, is the transfer of patients between different units within the same hospital.
Several studies have documented that risk of ICU death is highest among patients transferred from general medical‐surgical wards, intermediate among direct admissions from the emergency department, and lowest among surgical admissions.911 Opportunities to reduce subsequent ICU mortality have been studied among ward patients who develop sepsis and are then transferred to the ICU,12 among patients who experience cardiac arrest,13, 14 as well as among patients with any physiological deterioration (eg, through the use of rapid response teams).1517 Most of these studies have been single‐center studies and/or studies reporting only an ICU denominator. While useful in some respects, such studies are less helpful to hospitalists, who would benefit from better understanding of the types of patients transferred and the total impact that transfers to a higher level of care make on general medical‐surgical wards. In addition, entities such as the Institute for Healthcare Improvement recommend the manual review of records of patients who were transferred from the ward to the ICU18 to identify performance improvement opportunities. While laudable, such approaches do not lend themselves to automated reporting strategies.
We recently described a new risk adjustment methodology for inpatient mortality based entirely on automated data preceding hospital admission and not restricted to ICU patients. This methodology, which has been externally validated in Ottawa, Canada, after development in the Kaiser Permanente Medical Care Program (KPMCP), permits quantification of a patient's pre‐existing comorbidity burden, physiologic derangement at the time of admission, and overall inpatient mortality risk.19, 20 The primary purpose of this study was to combine this methodology with bed history analysis to quantify the in‐hospital mortality and length of stay (LOS) of patients who experienced intra‐hospital transfers in a large, multihospital system. As a secondary goal, we also wanted to assess the degree to which these transfers could be predicted based on information available prior to a patient's admission.
ABBREVIATIONS AND TERMS USED IN TEXT
COPS: COmorbidity Point Score. Point score based on a patient's health care utilization diagnoses (during the year preceding admission to the hospital. Analogous to POA (present on admission) coding. Scores can range from 0 to a theoretical maximum of 701 but scores >200 are rare. With respect to a patient's pre‐existing comorbidity burden, the unadjusted relationship of COPS and inpatient mortality is as follows: a COPS <50 is associated with a mortality risk of <1%, <100 with a mortality risk of <5%, 100 to 145 with a mortality risk of 5% to 10%, and >145 with a mortality risk of 10% or more.
ICU: Intensive Care Unit. In this study, all ICUs have a minimum registered nurse to patient ratio of 1:2.
LAPS: Laboratory Acute Physiology Score. Point score based on 14 laboratory test results obtained in the 72 hours preceding hospitalization. With respect to a patient's physiologic derangement, the unadjusted relationship of LAPS and inpatient mortality is as follows: a LAPS <7 is associated with a mortality risk of <1%, <7 to 30 with a mortality risk of <5%, 30 to 60 with a mortality risk of 5% to 9%, and >60 with a mortality risk of 10% or more.
LOS: Exact hospital Length Of Stay. LOS is calculated from admission until first discharge home (i.e., it may span more than one hospital stay if a patient experienced inter‐hospital transport).
Predicted (expected) mortality risk: the % risk of death for a given patient based on his/her age, sex, admission diagnosis, COPS, and LAPS.
OEMR: Observed to Expected Mortality Ratio. For a given patient subset, the ratio of the actual mortality experienced by the subset to the expected (predicted) mortality for the subset. Predicted mortality is based on patients' age, sex, admission diagnosis, COPS, and LAPS.
OMELOS: Observed Minus Expected LOS. For a given patient subset, the difference between the actual number of hospital days experienced by the subset and the expected (predicted) number of hospital days for the subset. Predicted LOS is based on patients' age, sex, admission diagnosis, COPS, and LAPS.
TCU: Transitional Care Unit (also called intermediate care unit or stepdown unit). In this study, TCUs have variable nurse to patient ratios ranging from 1:2.5 to 1:3 and did not provide assisted ventilation, continuous pressor infusions, or invasive monitoring.
Materials and Methods
This project was approved by the Northern California KPMCP Institutional Review Board for the Protection of Human Subjects.
The Northern California KPMCP serves a total population of approximately 3.3 million members. Under a mutual exclusivity arrangement, physicians of The Permanente Medical Group, Inc., care for Kaiser Foundation Health Plan, Inc. members at facilities owned by Kaiser Foundation Hospitals, Inc. All Northern California KPMCP hospitals and clinics employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere. Databases maintained by the KPMCP capture admission and discharge times, admission and discharge diagnoses and procedures (assigned by professional coders), bed histories, inter‐hospital transfers, as well as the results of all inpatient and outpatient laboratory tests. The use of these databases for research has been described in multiple reports.2124
Our setting consisted of all 19 hospitals owned and operated by the KPMCP, whose characteristics are summarized in the Supporting Information Appendix available to interested readers. These include the 17 described in our previous report19 as well as 2 new hospitals (Antioch and Manteca) which are similar in size and type of population served. Our study population consisted of all patients admitted to these 19 hospitals who met these criteria: 1) hospitalization began from November 1st, 2006 through January 31st, 2008; 2) initial hospitalization occurred at a Northern California KPMCP hospital (ie, for inter‐hospital transfers, the first hospital stay occurred within the KPMCP); 3) age 15 years; and 4) hospitalization was not for childbirth.
We defined a linked hospitalization as the time period that began with a patient's admission to the hospital and ended with the patient's discharge (home, to a nursing home, or death). Linked hospitalizations can thus involve more than 1 hospital stay and could include a patient transfer from one hospital to another prior to definitive discharge. For linked hospitalizations, mortality was attributed to the admitting KPMCP hospital (ie, if a patient was admitted to hospital A, transferred to B, and died at hospital B, mortality was attributed to hospital A). We defined total LOS as the exact time in hours from when a patient was first admitted to the hospital until death or final discharge home or to a nursing home, while total ICU or transitional care unit (TCU, referred to as stepdown unit in some hospitals) LOS was calculated for all individual ICU or TCU stays during the hospital stay.
Intra‐Hospital Transfers
We grouped all possible hospital units into four types: general medical‐surgical ward (henceforth, ward); operating room (OR)/post‐anesthesia recovery (PAR); TCU; and ICU. In 2003, the KPMCP implemented a mandatory minimum staffing ratio of one registered nurse for every four patients in all its hospital units; in addition, staffing levels for designated ICUs adhered to the previously mandated minimum of one nurse for every 2 patients. So long as they adhere to these minimum ratios, individual hospitals have considerable autonomy with respect to how they staff or designate individual hospital units. Registered nurse‐to‐patient ratios during the time of this study were as follows: ward patients, 1:3.5 to 1:4; TCU patients, 1:2.5 to 1:3; and ICU patients, 1:1 to 1:2. Staffing ratios for the OR and PAR are more variable, depending on the surgical procedures involved. Current KPMCP databases do not permit accurate quantification of physician staffing. All 19 study hospitals had designated ICUs, 6 were teaching hospitals, and 11 had designated TCUs. None of the study hospitals had closed ICUs (units where only intensivists admit patients) and none had continuous coverage of the ICU by intensivists. While we were not able to employ electronic data to determine who made the decision to transfer, we did find considerable variation with respect to how intensivists covered the ICUs and how they interfaced with hospitalists. Staffing levels for specialized coronary care units and non‐ICU monitored beds were not standardized. All study hospitals had rapid response teams as well as code blue teams during the time period covered by this report. Respiratory care practitioners were available to patients in all hospital units, but considerable variation existed with respect to other services available (eg, cardiac catheterization units, provision of noninvasive positive pressure ventilation outside the ICU, etc.).
This report focuses on intra‐hospital transfers to the ICU and TCU, with special emphasis on nonsurgical transfers (due to space limitations, we are not reporting on the outcomes of patients whose first hospital unit was the OR; additional details on these patients are provided in the Supporting Information Appendix). For the purposes of this report, we defined the following admission types: direct admits (patients admitted to the ICU or TCU whose first hospital unit on admission was the ICU or TCU); and nonsurgical transfers to a higher level of care. These latter transfers could be of 3 types: ward to ICU, ward to TCU, and TCU to ICU. We also quantified the effect of inter‐hospital transfers.
Independent Variables
In addition to patients' age and sex, we employed the following independent variables to predict transfer to a higher level of care. These variables are part of the risk adjustment model described in greater detail in our previous report19 and were available electronically for all patients in the cohort. We grouped admission diagnoses into 44 broad diagnostic categories (Primary Conditions), and admission types into 4 groups (emergency medical, emergency surgical, elective medical, and elective surgical). We quantified patients' degree of physiologic derangement using a Laboratory‐based Acute Physiology Score (LAPS) using laboratory test results prior to hospitalization. We quantified patients' comorbid illness burden using a Comorbidity Point Score (COPS) based on patients' pre‐existing diagnoses over the 12‐month period preceding hospitalization. Lastly, we assigned each patient a predicted mortality risk (%) and LOS based on the above predictors,19 permitting calculation of observed to expected mortality ratios (OEMRs) and observed minus expected LOS (OMELOS).
Statistical Methods
All analyses were performed in SAS.25 We calculated standard descriptive statistics (medians, means, standard deviations) and compared different patient groupings using t and chi‐square tests. We employed a similar approach to that reported by Render et al.7 to calculate OEMR and OMELOS.
To determine the degree to which transfers to a higher level of care from the ward or TCU would be predictable using information available at the time of admission, we performed 4 sets of logistic regression analyses using the above‐mentioned predictors in which the outcome variables were as follows: 1) transfer occurring in the first 48 hours after admission (time frame by which point approximately half of the transferred patients experienced a transfer) among ward or TCU patients and 2) transfer occurring after 48 hours among ward or TCU patients. We evaluated the discrimination and calibration of these models using the same methods described in our original report (measuring the area under the receiver operator characteristic curve, or c statistic, and visually examining observed and expected mortality rates among predicted risk bands as well as risk deciles) as well as additional statistical tests recommended by Cook.19, 26
Results
During the study period, a total of 249,129 individual hospital stays involving 170,151 patients occurred at these 19 hospitals. After concatenation of inter‐hospital transfers, we were left with 237,208 linked hospitalizations. We excluded 26,738 linked hospitalizations that began at a non‐KPMCP hospital (ie, they were transported in), leaving a total of 210,470 linked hospitalizations involving 150,495 patients. The overall linked hospitalization mortality rate was 3.30%.
Table 1 summarizes cohort characteristics based on initial hospital location. On admission, ICU patients had the highest degree of physiologic derangement as well as the highest predicted mortality. Considerable inter‐hospital variation was present in both predictors and outcomes; details on these variations are provided in the Supporting Information Appendix.
Ward | TCU | ICU | All* | |
---|---|---|---|---|
| ||||
n | 121,237 | 20,556 | 16,001 | 210,470 |
Admitted via emergency department, n (%) | 99,909 (82.4) | 18,612 (90.5) | 13,847 (86.5) | 139,036 (66.1) |
% range across hospitals | 55.0‐94.2 | 64.7‐97.6 | 49.5‐97.4 | 53.6‐76.9 |
Male, n (%) | 53,744 (44.3) | 10,362 (50.4) | 8,378 (52.4) | 94,451 (44.9) |
Age in years (mean SD) | 64.5 19.2 | 69.0 15.6 | 63.7 17.8 | 63.2 18.6 |
LAPS (mean SD) | 19.2 18.0 | 23.3 19.5 | 31.7 25.7 | 16.7 19.0 |
COPS (mean SD) | 90.4 64.0 | 99.2 65.9 | 94.5 67.5 | 84.7 61.8 |
% predicted mortality (mean SD) | 4.0 7.1 | 4.6 7.3 | 8.7 12.8 | 3.6 7.3 |
Observed in‐hospital deaths (n, %) | 3,793 (3.1) | 907 (4.4) | 1,995 (12.5) | 6,952 (3.3) |
Observed to expected mortality ratio | 0.79 (0.77‐0.82) | 0.95 (0.89‐1.02) | 1.43 (1.36‐1.49) | 0.92 (0.89‐0.94) |
Total hospital LOS, days (mean SD) | 4.6 7.5 | 5.3 10.0 | 7.8 14.0 | 4.6 8.1 |
Table 2 summarizes data from 3 groups of patients: patients initially admitted to the ward, or TCU, who did not experience a transfer to a higher level of care and patients admitted to these 2 units who did experience such a transfer. Patients who experienced a transfer constituted 5.3% (6,484/121,237) of ward patients and 6.7% (1,384/20,556) of TCU patients. Transferred patients tended to be older, have more acute physiologic derangement (higher LAPS), a greater pre‐existing illness burden (higher COPS), and a higher predicted mortality risk. Among ward patients, those with the following admission diagnoses were most likely to experience a transfer to a higher level of care: gastrointestinal bleeding (10.8% of all transfers), pneumonia (8.7%), and other infections (8.2%). The diagnoses most likely to be associated with death following transfer were cancer (death rate among transferred patients, 48%), renal disease (death rate, 36%), and liver disease (33%). Similar distributions were observed for TCU patients.
Patients Initially Admitted to Ward, Remained There | Patients Initially Admitted to TCU, Remained There | Patients Transferred to Higher Level of Care | All | |
---|---|---|---|---|
| ||||
n | 114,753 | 19,172 | 7,868 | 141,793 |
Male, n (%) | 50,586 (44.1) | 9,626 (50.2) | 3,894 (49.5) | 64,106 (45.2) |
Age (mean SD) | 64.3 19.4 | 69.0 15.7 | 68.1 16.1 | 65.2 18.8 |
LAPS (mean SD) | 18.9 17.8 | 22.7 19.1 | 26.7 21.0 | 19.8 18.3 |
COPS (mean SD) | 89.4 63.7 | 98.3 65.5 | 107.9 67.6 | 91.7 64.4 |
% predicted mortality risk (mean SD) | 3.8 7.0 | 4.4 7.0 | 6.5 8.8 | 4.1 7.1 |
Admission diagnosis of pneumonia, n (%) | 5,624 (4.9) | 865 (4.5) | 684 (8.7) | 7,173 (5.1) |
Admission diagnosis of sepsis, n (%) | 1,181 (1.0) | 227 (1.2) | 168 (2.1) | 1,576 (1.1) |
Admission diagnosis of GI bleed, n (%) | 13,615 (11.9) | 1,448 (7.6) | 851 (10.8) | 15,914 (11.2) |
Admission diagnosis of cancer, n (%) | 2,406 (2.1) | 80 (0.4) | 186 (2.4) | 2,672 (1.9) |
Table 3 compares outcomes among ward and TCU patients who did and did not experience a transfer to a higher level of care. The table shows that transferred patients were almost 3 times as likely to die, even after controlling for severity of illness, and that their hospital LOS was 9 days higher than expected. This increased risk was seen in all hospitals and among all transfer types (ward to ICU, ward to TCU, and TCU to ICU).
Patients Initially Admitted to Ward, Remained There | Patients Initially Admitted to TCU, Remained There | Patients Transferred to Higher Level of Care | |
---|---|---|---|
| |||
n | 114,753 | 19,172 | 7,868 |
Admitted to ICU, n (%) | 0 (0.0) | 0 (0.0) | 5,245 (66.7) |
Ventilated, n (%) | 0 (0.0) | 0 (0.0) | 1,346 (17.1) |
Died in the hospital, n (%) | 2,619 (2.3) | 572 (3.0) | 1,509 (19.2) |
Length of stay, in days, at time of death (mean SD) | 7.0 11.9 | 8.3 12.4 | 16.2 23.7 |
Observed to expected mortality ratio (95% CI) | 0.60 (0.57‐0.62) | 0.68 (0.63‐0.74) | 2.93 (2.79‐3.09) |
Total hospital length of stay, days (mean SD) | 4.0 5.7 | 4.4 6.9 | 14.3 21.3 |
Observed minus expected length of stay (95% CI) | 0.4 (0.3‐0.4) | 0.8 (0.7‐0.9) | 9.1 (8.6‐9.5) |
Length of stay, in hours, at time of transfer (mean SD) | 80.8 167.2 |
Table 3 also shows that, among decedent patients, those who never left the ward or TCU died much sooner than those who died following transfer. Among direct admits to the ICU, the median LOS at time of death was 3.9 days, with a mean of 9.4 standard deviation of 19.9 days, while the corresponding times for TCU direct admits were a median and mean LOS of 6.5 and 11.7 19.5 days.
Table 4 summarizes outcomes among different patient subgroups that did and did not experience a transfer to a higher level of care. Based on location, patients who experienced a transfer from the TCU to the ICU had the highest crude death rate, but patients transferred from the ward to the ICU had the highest OEMR. On the other hand, if one divides patients by the degree of physiologic derangement, patients with low LAPS who experienced a transfer had the highest OEMR. With respect to LOS, patients transferred from the TCU to the ICU had the highest OMELOS (13.4 extra days).
n (%)* | Death Rate (%) | OEMR | LOS (mean SD) | OMELOS | |
---|---|---|---|---|---|
| |||||
Never admitted to TCU or ICU | 157,632 (74.9) | 1.6 | 0.55 (0.53‐0.57) | 3.6 4.6 | 0.04 (0.02‐0.07) |
Direct admit to TCU | 18,464 (8.8) | 2.9 | 0.66 (0.61‐0.72) | 4.2 5.8 | 0.60 (0.52‐0.68) |
Direct admit to ICU | 14,655 (7.0) | 11.9 | 1.38 (1.32‐1.45) | 6.4 9.4 | 2.28 (2.14‐2.43) |
Transferred from ward to ICU | 5,145 (2.4) | 21.5 | 3.23 (3.04‐3.42) | 15.7 21.6 | 10.33 (9.70‐10.96) |
Transferred from ward to TCU | 3,144 (1.5) | 11.9 | 1.99 (1.79‐2.20) | 13.6 23.2 | 8.02 (7.23‐8.82) |
Transferred from TCU to ICU | 1,107 (0.5) | 25.7 | 2.94 (2.61‐3.31) | 18.0 28.2 | 13.35 (11.49‐15.21) |
Admitted to ward, COPS 80, no transfer to ICU or TCU | 55,405 (26.3) | 3.4 | 0.59 (0.56‐0.62) | 4.5 5.9 | 0.29 (0.24‐0.34) |
Admitted to ward, COPS 80, did experience transfer to ICU or TCU | 4,851 (2.3) | 19.3 | 2.72 (2.55‐2.90) | 14.2 20.0 | 8.14 (7.56‐8.71) |
Admitted to ward, COPS <80, no transfer to ICU or TCU | 57,421 (27.3) | 1.1 | 0.55 (0.51‐0.59) | 3.4 4.2 | 0.23 (0.19‐0.26) |
Admitted to ward, COPS <80, did experience transfer to ICU or TCU | 3,560 (1.7) | 9.8 | 2.93 (2.63‐3.26) | 12.0 19.0 | 7.52 (6.89‐8.15) |
Admitted to ward, LAPS 20, no transfer to ICU or TCU | 46,492 (22.1) | 4.2 | 0.59 (0.56‐0.61) | 4.6 5.4 | 0.16 (0.12‐0.21) |
Admitted to ward, LAPS 20, did experience transfer to ICU or TCU | 4,070 (1.9) | 21.4 | 2.37 (2.22‐2.54) | 14.8 21.0 | 8.76 (8.06‐9.47) |
Admitted to ward, LAPS <20, no transfer to ICU or TCU | 66,334 (31.5) | 0.9 | 0.55 (0.51‐0.60) | 3.5 4.9 | 0.32 (0.28‐0.36) |
Admitted to ward, LAPS <20, did experience transfer to ICU or TCU | 4,341 (2.1) | 9.5 | 4.31 (3.90‐4.74) | 11.8 18.1 | 7.12 (6.61‐7.64) |
Transfers to a higher level of care at a different hospital, which in the KPMCP are usually planned, experienced lower mortality than transfers within the same hospital. For ward to TCU transfers, intra‐hospital transfers had a mortality of 12.1% while inter‐hospital transfers had a mortality of 5.7%. Corresponding rates for ward to ICU transfers were 21.7% and 11.2%, and for TCU to ICU transfers the rates were 25.9% and 12.5%, respectively.
Among patients initially admitted to the ward, a model to predict the occurrence of a transfer to a higher level of care (within 48 hours after admission) that included age, sex, admission type, primary condition, LAPS, COPS, and interaction terms had poor discrimination, with an area under the receiver operator characteristic (c statistic) of only 0.64. The c statistic for a model to predict transfer after 48 hours was 0.66. The corresponding models for TCU admits had c statistics of 0.67 and 0.68. All four models had poor calibration.
Discussion
Using automated bed history data permits characterizing a patient population with disproportionate mortality and LOS: intra‐hospital transfers to special care units (ICUs or TCUs). Indeed, the largest subset of these patients (those initially admitted to the ward or TCU) constituted only 3.7% of all admissions, but accounted for 24.2% of all ICU admissions, 21.7% of all hospital deaths, and 13.2% of all hospital days. These patients also had very elevated OEMRs and OMELOS. Models based on age, sex, preadmission laboratory test results, and comorbidities did not predict the occurrence of these transfers.
We performed multivariate analyses to explore the degree to which electronically assigned preadmission severity scores could predict these transfers. These analyses found that, compared to our ability to predict inpatient or 30‐day mortality at the time of admission, which is excellent, our ability to predict the occurrence of transfer after admission is much more limited. These results highlight the limitations of severity scores that rely on automated data, which may not have adequate discrimination when it comes to determining the risk of an adverse outcome within a narrow time frame. For example, among the 121,237 patients initially admitted to the ward who did not experience an intra‐hospital transfer, the mean LAPS was 18.9, while the mean LAPS among the 6,484 ward patients who did experience a transfer was 25.5. Differences between the mean and median LAPS, COPS, and predicted mortality risk among transferred and non‐transferred patients were significant (P < 0.0001 for all comparisons). However, examination of the distribution of LAPS, COPS, and predicted mortality risk between these two groups of patients showed considerable overlap.
Our methodology resembles Silber et al.'s27, 28 concept of failure to rescue in that it focuses on events occurring after hospitalization. Silber et al. argue that a hospital's quality can be measured by quantifying the degree to which patients who experience new problems are successfully rescued. Furthermore, quantification of those situations where rescue attempts are unsuccessful is felt to be superior to simply comparing raw or adjusted mortality rates because these are primarily determined by underlying case mix. The primary difference between Silber et al.'s approach and ours is at the level of detailthey specified a specific set of complications, whereas our measure is more generic and would include patients with many of the complications specified by Silber et al.27, 28
Most of the patients transferred to a higher level of care in our cohort survived (ie, were rescued), indicating that intensive care is beneficial. However, the fact that these patients had elevated OEMRs and OMELOS indicates that the real challenge facing hospitalists involves the timing of provision of a beneficial intervention. In theory, improved timing could result from earlier detection of problems, which is the underlying rationale for employing rapid response teams. However, the fact that our electronic tools (LAPS, COPS) cannot predict patient deteriorations within a narrow time frame suggests that early detection will remain a major challenge. Manually assigned vital signs scores designed for this purpose do not have good discrimination either.29, 30 This raises the possibility that, though patient groups may differ in terms of overall illness severity and mortality risk, differences at the individual patient level may be too subtle for clinicians to detect. Future research may thus need to focus on scores that combine laboratory data, vital signs, trends in data,31, 32 and newer proteomic markers (eg, procalcitonin).33 We also found that most transfers occurred early (within <72 hours), raising the possibility that at least some of these transfers may involve issues around triage rather than sudden deterioration.
Our study has important limitations. Due to resource constraints and limited data availability, we could not characterize the patients as well as might be desirable; in particular, we could not make full determinations of the actual reasons for patients' transfer for all patients. Broadly speaking, transfer to a higher level of care could be due to inappropriate triage, appropriate (preventive) transfer (which could include transfer to a more richly staffed unit for a specific procedure), relentless progression of disease despite maximal therapy, the occurrence of management errors, patient and family uncertainty about goals of care or inadequate understanding of treatment options and prognoses, or a combination of these factors. We could not make these distinctions with currently available electronic data. This is also true of postsurgical patients, in whom it is difficult to determine which transfers to intensive care might be planned (eg, in the case of surgical procedures where ICU care is anticipated) as opposed to the occurrence of a deterioration during or following surgery. Another major limitation of this study is our inability to identify code or no code status electronically. The elapsed LOS at time of death among patients who experienced a transfer to a higher level of care (as compared to patients who died in the ward without ever experiencing intra‐hospital transfer) suggests, but does not prove, that prolonged efforts were being made to keep them alive. We were also limited in terms of having access to other process data (eg, physician staffing levels, provision and timing of palliative care). Having ICU severity of illness scores would have permitted us to compare our cohort to those of other recent studies showing elevated mortality rates among transfer patients,911 but we have not yet developed that capability.
Consideration of our study findings suggests a possible research agenda that could be implemented by hospitalist researchers. This agenda should emphasize three areas: detection, intervention, and reflection.
With respect to detection, attention needs to be paid to better tools for quantifying patient risk at the time a decision to admit to the ward is made. It is likely that such tools will need to combine the attributes of our severity score (LAPS) with those of the manually assigned scores.30, 34 In some cases, use of these tools could lead a physician to change the locus of admission from the ward to the TCU or ICU, which could improve outcomes by ensuring more timely provision of intensive care. Since problems with initial triage could be due to factors other than the failure to suspect or anticipate impending instability, future research should also include a cognitive component (eg, quantifying what proportion of subsequent patient deteriorations could be ascribed to missed diagnoses35). Additional work also needs to be done on developing mathematical models that can inform electronic monitoring of ward (not just ICU) patients.
Research on interventions that hospitalists can use to prevent the need for intensive care or to improve the rescue rate should take two routes. The first is a disease‐specific route, which builds on the fact that a relatively small set of conditions (pneumonia, sepsis, gastrointestinal bleeding) account for most transfers to a higher level of care. Condition‐specific protocols, checklists, and bundles36 tailored to a ward environment (as opposed to the ICU or to the entire hospital) might prevent deteriorations in these patients, as has been reported for sepsis.37 The second route is to improve the overall capabilities of rapid response and code blue teams. Such research would need to include a more careful assessment of what commonalities exist among patients who were and were not successfully rescued by these teams. This approach would probably yield more insights than the current literature, which focuses on whether rapid response teams are a good thing or not.
Finally, research also needs to be performed on how hospitalists reflect on adverse outcomes among ward patients. Greater emphasis needs to be placed on moving beyond trigger tool approaches that rely on manual chart review. In an era of expanding use of electronic medical record systems, more work needs to be done on how to harness these to provide hospitalists with better quantitative and risk‐adjusted information. This information should not be limited to simply reporting rates of transfers and deaths. Rather, finer distinctions must be provided with respect of the type of patients (ie, more diagnostic detail), the clinical status of patients (ie, more physiologic detail), as well as the effects of including or excluding patients in whom therapeutic options may be limited (ie, do not resuscitate and comfort care patients) on reported rates. Ideally, researchers should develop better process and outcomes measures that could be tested in collaborative networks that include multiple nonacademic general medical‐surgical wards.
Acknowledgements
The authors thank Drs. Paul Feigenbaum, Alan Whippy, Joseph V. Selby, and Philip Madvig for reviewing the manuscript and Ms. Jennifer Calhoun for formatting the manuscript.
Considerable research and public attention is being paid to the quantification, risk adjustment, and reporting of inpatient mortality.15 Inpatient mortality is reported as aggregate mortality (for all hospitalized patients or those with a specific diagnosis3, 6) or intensive care unit (ICU) mortality.7, 8 While reporting aggregate hospital or aggregate ICU mortality rates is useful, it is also important to develop reporting strategies that go beyond simply using data elements found in administrative databases (eg, diagnosis and procedure codes) to quantify practice variation. Ideally, such strategies would permit delineating processes of careparticularly those potentially under the control of hospitalists, not only intensiviststo identify improvement opportunities. One such process, which can be tracked using the bed history component of a patient's electronic medical record, is the transfer of patients between different units within the same hospital.
Several studies have documented that risk of ICU death is highest among patients transferred from general medical‐surgical wards, intermediate among direct admissions from the emergency department, and lowest among surgical admissions.911 Opportunities to reduce subsequent ICU mortality have been studied among ward patients who develop sepsis and are then transferred to the ICU,12 among patients who experience cardiac arrest,13, 14 as well as among patients with any physiological deterioration (eg, through the use of rapid response teams).1517 Most of these studies have been single‐center studies and/or studies reporting only an ICU denominator. While useful in some respects, such studies are less helpful to hospitalists, who would benefit from better understanding of the types of patients transferred and the total impact that transfers to a higher level of care make on general medical‐surgical wards. In addition, entities such as the Institute for Healthcare Improvement recommend the manual review of records of patients who were transferred from the ward to the ICU18 to identify performance improvement opportunities. While laudable, such approaches do not lend themselves to automated reporting strategies.
We recently described a new risk adjustment methodology for inpatient mortality based entirely on automated data preceding hospital admission and not restricted to ICU patients. This methodology, which has been externally validated in Ottawa, Canada, after development in the Kaiser Permanente Medical Care Program (KPMCP), permits quantification of a patient's pre‐existing comorbidity burden, physiologic derangement at the time of admission, and overall inpatient mortality risk.19, 20 The primary purpose of this study was to combine this methodology with bed history analysis to quantify the in‐hospital mortality and length of stay (LOS) of patients who experienced intra‐hospital transfers in a large, multihospital system. As a secondary goal, we also wanted to assess the degree to which these transfers could be predicted based on information available prior to a patient's admission.
ABBREVIATIONS AND TERMS USED IN TEXT
COPS: COmorbidity Point Score. Point score based on a patient's health care utilization diagnoses (during the year preceding admission to the hospital. Analogous to POA (present on admission) coding. Scores can range from 0 to a theoretical maximum of 701 but scores >200 are rare. With respect to a patient's pre‐existing comorbidity burden, the unadjusted relationship of COPS and inpatient mortality is as follows: a COPS <50 is associated with a mortality risk of <1%, <100 with a mortality risk of <5%, 100 to 145 with a mortality risk of 5% to 10%, and >145 with a mortality risk of 10% or more.
ICU: Intensive Care Unit. In this study, all ICUs have a minimum registered nurse to patient ratio of 1:2.
LAPS: Laboratory Acute Physiology Score. Point score based on 14 laboratory test results obtained in the 72 hours preceding hospitalization. With respect to a patient's physiologic derangement, the unadjusted relationship of LAPS and inpatient mortality is as follows: a LAPS <7 is associated with a mortality risk of <1%, <7 to 30 with a mortality risk of <5%, 30 to 60 with a mortality risk of 5% to 9%, and >60 with a mortality risk of 10% or more.
LOS: Exact hospital Length Of Stay. LOS is calculated from admission until first discharge home (i.e., it may span more than one hospital stay if a patient experienced inter‐hospital transport).
Predicted (expected) mortality risk: the % risk of death for a given patient based on his/her age, sex, admission diagnosis, COPS, and LAPS.
OEMR: Observed to Expected Mortality Ratio. For a given patient subset, the ratio of the actual mortality experienced by the subset to the expected (predicted) mortality for the subset. Predicted mortality is based on patients' age, sex, admission diagnosis, COPS, and LAPS.
OMELOS: Observed Minus Expected LOS. For a given patient subset, the difference between the actual number of hospital days experienced by the subset and the expected (predicted) number of hospital days for the subset. Predicted LOS is based on patients' age, sex, admission diagnosis, COPS, and LAPS.
TCU: Transitional Care Unit (also called intermediate care unit or stepdown unit). In this study, TCUs have variable nurse to patient ratios ranging from 1:2.5 to 1:3 and did not provide assisted ventilation, continuous pressor infusions, or invasive monitoring.
Materials and Methods
This project was approved by the Northern California KPMCP Institutional Review Board for the Protection of Human Subjects.
The Northern California KPMCP serves a total population of approximately 3.3 million members. Under a mutual exclusivity arrangement, physicians of The Permanente Medical Group, Inc., care for Kaiser Foundation Health Plan, Inc. members at facilities owned by Kaiser Foundation Hospitals, Inc. All Northern California KPMCP hospitals and clinics employ the same information systems with a common medical record number and can track care covered by the plan but delivered elsewhere. Databases maintained by the KPMCP capture admission and discharge times, admission and discharge diagnoses and procedures (assigned by professional coders), bed histories, inter‐hospital transfers, as well as the results of all inpatient and outpatient laboratory tests. The use of these databases for research has been described in multiple reports.2124
Our setting consisted of all 19 hospitals owned and operated by the KPMCP, whose characteristics are summarized in the Supporting Information Appendix available to interested readers. These include the 17 described in our previous report19 as well as 2 new hospitals (Antioch and Manteca) which are similar in size and type of population served. Our study population consisted of all patients admitted to these 19 hospitals who met these criteria: 1) hospitalization began from November 1st, 2006 through January 31st, 2008; 2) initial hospitalization occurred at a Northern California KPMCP hospital (ie, for inter‐hospital transfers, the first hospital stay occurred within the KPMCP); 3) age 15 years; and 4) hospitalization was not for childbirth.
We defined a linked hospitalization as the time period that began with a patient's admission to the hospital and ended with the patient's discharge (home, to a nursing home, or death). Linked hospitalizations can thus involve more than 1 hospital stay and could include a patient transfer from one hospital to another prior to definitive discharge. For linked hospitalizations, mortality was attributed to the admitting KPMCP hospital (ie, if a patient was admitted to hospital A, transferred to B, and died at hospital B, mortality was attributed to hospital A). We defined total LOS as the exact time in hours from when a patient was first admitted to the hospital until death or final discharge home or to a nursing home, while total ICU or transitional care unit (TCU, referred to as stepdown unit in some hospitals) LOS was calculated for all individual ICU or TCU stays during the hospital stay.
Intra‐Hospital Transfers
We grouped all possible hospital units into four types: general medical‐surgical ward (henceforth, ward); operating room (OR)/post‐anesthesia recovery (PAR); TCU; and ICU. In 2003, the KPMCP implemented a mandatory minimum staffing ratio of one registered nurse for every four patients in all its hospital units; in addition, staffing levels for designated ICUs adhered to the previously mandated minimum of one nurse for every 2 patients. So long as they adhere to these minimum ratios, individual hospitals have considerable autonomy with respect to how they staff or designate individual hospital units. Registered nurse‐to‐patient ratios during the time of this study were as follows: ward patients, 1:3.5 to 1:4; TCU patients, 1:2.5 to 1:3; and ICU patients, 1:1 to 1:2. Staffing ratios for the OR and PAR are more variable, depending on the surgical procedures involved. Current KPMCP databases do not permit accurate quantification of physician staffing. All 19 study hospitals had designated ICUs, 6 were teaching hospitals, and 11 had designated TCUs. None of the study hospitals had closed ICUs (units where only intensivists admit patients) and none had continuous coverage of the ICU by intensivists. While we were not able to employ electronic data to determine who made the decision to transfer, we did find considerable variation with respect to how intensivists covered the ICUs and how they interfaced with hospitalists. Staffing levels for specialized coronary care units and non‐ICU monitored beds were not standardized. All study hospitals had rapid response teams as well as code blue teams during the time period covered by this report. Respiratory care practitioners were available to patients in all hospital units, but considerable variation existed with respect to other services available (eg, cardiac catheterization units, provision of noninvasive positive pressure ventilation outside the ICU, etc.).
This report focuses on intra‐hospital transfers to the ICU and TCU, with special emphasis on nonsurgical transfers (due to space limitations, we are not reporting on the outcomes of patients whose first hospital unit was the OR; additional details on these patients are provided in the Supporting Information Appendix). For the purposes of this report, we defined the following admission types: direct admits (patients admitted to the ICU or TCU whose first hospital unit on admission was the ICU or TCU); and nonsurgical transfers to a higher level of care. These latter transfers could be of 3 types: ward to ICU, ward to TCU, and TCU to ICU. We also quantified the effect of inter‐hospital transfers.
Independent Variables
In addition to patients' age and sex, we employed the following independent variables to predict transfer to a higher level of care. These variables are part of the risk adjustment model described in greater detail in our previous report19 and were available electronically for all patients in the cohort. We grouped admission diagnoses into 44 broad diagnostic categories (Primary Conditions), and admission types into 4 groups (emergency medical, emergency surgical, elective medical, and elective surgical). We quantified patients' degree of physiologic derangement using a Laboratory‐based Acute Physiology Score (LAPS) using laboratory test results prior to hospitalization. We quantified patients' comorbid illness burden using a Comorbidity Point Score (COPS) based on patients' pre‐existing diagnoses over the 12‐month period preceding hospitalization. Lastly, we assigned each patient a predicted mortality risk (%) and LOS based on the above predictors,19 permitting calculation of observed to expected mortality ratios (OEMRs) and observed minus expected LOS (OMELOS).
Statistical Methods
All analyses were performed in SAS.25 We calculated standard descriptive statistics (medians, means, standard deviations) and compared different patient groupings using t and chi‐square tests. We employed a similar approach to that reported by Render et al.7 to calculate OEMR and OMELOS.
To determine the degree to which transfers to a higher level of care from the ward or TCU would be predictable using information available at the time of admission, we performed 4 sets of logistic regression analyses using the above‐mentioned predictors in which the outcome variables were as follows: 1) transfer occurring in the first 48 hours after admission (time frame by which point approximately half of the transferred patients experienced a transfer) among ward or TCU patients and 2) transfer occurring after 48 hours among ward or TCU patients. We evaluated the discrimination and calibration of these models using the same methods described in our original report (measuring the area under the receiver operator characteristic curve, or c statistic, and visually examining observed and expected mortality rates among predicted risk bands as well as risk deciles) as well as additional statistical tests recommended by Cook.19, 26
Results
During the study period, a total of 249,129 individual hospital stays involving 170,151 patients occurred at these 19 hospitals. After concatenation of inter‐hospital transfers, we were left with 237,208 linked hospitalizations. We excluded 26,738 linked hospitalizations that began at a non‐KPMCP hospital (ie, they were transported in), leaving a total of 210,470 linked hospitalizations involving 150,495 patients. The overall linked hospitalization mortality rate was 3.30%.
Table 1 summarizes cohort characteristics based on initial hospital location. On admission, ICU patients had the highest degree of physiologic derangement as well as the highest predicted mortality. Considerable inter‐hospital variation was present in both predictors and outcomes; details on these variations are provided in the Supporting Information Appendix.
Ward | TCU | ICU | All* | |
---|---|---|---|---|
| ||||
n | 121,237 | 20,556 | 16,001 | 210,470 |
Admitted via emergency department, n (%) | 99,909 (82.4) | 18,612 (90.5) | 13,847 (86.5) | 139,036 (66.1) |
% range across hospitals | 55.0‐94.2 | 64.7‐97.6 | 49.5‐97.4 | 53.6‐76.9 |
Male, n (%) | 53,744 (44.3) | 10,362 (50.4) | 8,378 (52.4) | 94,451 (44.9) |
Age in years (mean SD) | 64.5 19.2 | 69.0 15.6 | 63.7 17.8 | 63.2 18.6 |
LAPS (mean SD) | 19.2 18.0 | 23.3 19.5 | 31.7 25.7 | 16.7 19.0 |
COPS (mean SD) | 90.4 64.0 | 99.2 65.9 | 94.5 67.5 | 84.7 61.8 |
% predicted mortality (mean SD) | 4.0 7.1 | 4.6 7.3 | 8.7 12.8 | 3.6 7.3 |
Observed in‐hospital deaths (n, %) | 3,793 (3.1) | 907 (4.4) | 1,995 (12.5) | 6,952 (3.3) |
Observed to expected mortality ratio | 0.79 (0.77‐0.82) | 0.95 (0.89‐1.02) | 1.43 (1.36‐1.49) | 0.92 (0.89‐0.94) |
Total hospital LOS, days (mean SD) | 4.6 7.5 | 5.3 10.0 | 7.8 14.0 | 4.6 8.1 |
Table 2 summarizes data from 3 groups of patients: patients initially admitted to the ward, or TCU, who did not experience a transfer to a higher level of care and patients admitted to these 2 units who did experience such a transfer. Patients who experienced a transfer constituted 5.3% (6,484/121,237) of ward patients and 6.7% (1,384/20,556) of TCU patients. Transferred patients tended to be older, have more acute physiologic derangement (higher LAPS), a greater pre‐existing illness burden (higher COPS), and a higher predicted mortality risk. Among ward patients, those with the following admission diagnoses were most likely to experience a transfer to a higher level of care: gastrointestinal bleeding (10.8% of all transfers), pneumonia (8.7%), and other infections (8.2%). The diagnoses most likely to be associated with death following transfer were cancer (death rate among transferred patients, 48%), renal disease (death rate, 36%), and liver disease (33%). Similar distributions were observed for TCU patients.
Patients Initially Admitted to Ward, Remained There | Patients Initially Admitted to TCU, Remained There | Patients Transferred to Higher Level of Care | All | |
---|---|---|---|---|
| ||||
n | 114,753 | 19,172 | 7,868 | 141,793 |
Male, n (%) | 50,586 (44.1) | 9,626 (50.2) | 3,894 (49.5) | 64,106 (45.2) |
Age (mean SD) | 64.3 19.4 | 69.0 15.7 | 68.1 16.1 | 65.2 18.8 |
LAPS (mean SD) | 18.9 17.8 | 22.7 19.1 | 26.7 21.0 | 19.8 18.3 |
COPS (mean SD) | 89.4 63.7 | 98.3 65.5 | 107.9 67.6 | 91.7 64.4 |
% predicted mortality risk (mean SD) | 3.8 7.0 | 4.4 7.0 | 6.5 8.8 | 4.1 7.1 |
Admission diagnosis of pneumonia, n (%) | 5,624 (4.9) | 865 (4.5) | 684 (8.7) | 7,173 (5.1) |
Admission diagnosis of sepsis, n (%) | 1,181 (1.0) | 227 (1.2) | 168 (2.1) | 1,576 (1.1) |
Admission diagnosis of GI bleed, n (%) | 13,615 (11.9) | 1,448 (7.6) | 851 (10.8) | 15,914 (11.2) |
Admission diagnosis of cancer, n (%) | 2,406 (2.1) | 80 (0.4) | 186 (2.4) | 2,672 (1.9) |
Table 3 compares outcomes among ward and TCU patients who did and did not experience a transfer to a higher level of care. The table shows that transferred patients were almost 3 times as likely to die, even after controlling for severity of illness, and that their hospital LOS was 9 days higher than expected. This increased risk was seen in all hospitals and among all transfer types (ward to ICU, ward to TCU, and TCU to ICU).
Patients Initially Admitted to Ward, Remained There | Patients Initially Admitted to TCU, Remained There | Patients Transferred to Higher Level of Care | |
---|---|---|---|
| |||
n | 114,753 | 19,172 | 7,868 |
Admitted to ICU, n (%) | 0 (0.0) | 0 (0.0) | 5,245 (66.7) |
Ventilated, n (%) | 0 (0.0) | 0 (0.0) | 1,346 (17.1) |
Died in the hospital, n (%) | 2,619 (2.3) | 572 (3.0) | 1,509 (19.2) |
Length of stay, in days, at time of death (mean SD) | 7.0 11.9 | 8.3 12.4 | 16.2 23.7 |
Observed to expected mortality ratio (95% CI) | 0.60 (0.57‐0.62) | 0.68 (0.63‐0.74) | 2.93 (2.79‐3.09) |
Total hospital length of stay, days (mean SD) | 4.0 5.7 | 4.4 6.9 | 14.3 21.3 |
Observed minus expected length of stay (95% CI) | 0.4 (0.3‐0.4) | 0.8 (0.7‐0.9) | 9.1 (8.6‐9.5) |
Length of stay, in hours, at time of transfer (mean SD) | 80.8 167.2 |
Table 3 also shows that, among decedent patients, those who never left the ward or TCU died much sooner than those who died following transfer. Among direct admits to the ICU, the median LOS at time of death was 3.9 days, with a mean of 9.4 standard deviation of 19.9 days, while the corresponding times for TCU direct admits were a median and mean LOS of 6.5 and 11.7 19.5 days.
Table 4 summarizes outcomes among different patient subgroups that did and did not experience a transfer to a higher level of care. Based on location, patients who experienced a transfer from the TCU to the ICU had the highest crude death rate, but patients transferred from the ward to the ICU had the highest OEMR. On the other hand, if one divides patients by the degree of physiologic derangement, patients with low LAPS who experienced a transfer had the highest OEMR. With respect to LOS, patients transferred from the TCU to the ICU had the highest OMELOS (13.4 extra days).
n (%)* | Death Rate (%) | OEMR | LOS (mean SD) | OMELOS | |
---|---|---|---|---|---|
| |||||
Never admitted to TCU or ICU | 157,632 (74.9) | 1.6 | 0.55 (0.53‐0.57) | 3.6 4.6 | 0.04 (0.02‐0.07) |
Direct admit to TCU | 18,464 (8.8) | 2.9 | 0.66 (0.61‐0.72) | 4.2 5.8 | 0.60 (0.52‐0.68) |
Direct admit to ICU | 14,655 (7.0) | 11.9 | 1.38 (1.32‐1.45) | 6.4 9.4 | 2.28 (2.14‐2.43) |
Transferred from ward to ICU | 5,145 (2.4) | 21.5 | 3.23 (3.04‐3.42) | 15.7 21.6 | 10.33 (9.70‐10.96) |
Transferred from ward to TCU | 3,144 (1.5) | 11.9 | 1.99 (1.79‐2.20) | 13.6 23.2 | 8.02 (7.23‐8.82) |
Transferred from TCU to ICU | 1,107 (0.5) | 25.7 | 2.94 (2.61‐3.31) | 18.0 28.2 | 13.35 (11.49‐15.21) |
Admitted to ward, COPS 80, no transfer to ICU or TCU | 55,405 (26.3) | 3.4 | 0.59 (0.56‐0.62) | 4.5 5.9 | 0.29 (0.24‐0.34) |
Admitted to ward, COPS 80, did experience transfer to ICU or TCU | 4,851 (2.3) | 19.3 | 2.72 (2.55‐2.90) | 14.2 20.0 | 8.14 (7.56‐8.71) |
Admitted to ward, COPS <80, no transfer to ICU or TCU | 57,421 (27.3) | 1.1 | 0.55 (0.51‐0.59) | 3.4 4.2 | 0.23 (0.19‐0.26) |
Admitted to ward, COPS <80, did experience transfer to ICU or TCU | 3,560 (1.7) | 9.8 | 2.93 (2.63‐3.26) | 12.0 19.0 | 7.52 (6.89‐8.15) |
Admitted to ward, LAPS 20, no transfer to ICU or TCU | 46,492 (22.1) | 4.2 | 0.59 (0.56‐0.61) | 4.6 5.4 | 0.16 (0.12‐0.21) |
Admitted to ward, LAPS 20, did experience transfer to ICU or TCU | 4,070 (1.9) | 21.4 | 2.37 (2.22‐2.54) | 14.8 21.0 | 8.76 (8.06‐9.47) |
Admitted to ward, LAPS <20, no transfer to ICU or TCU | 66,334 (31.5) | 0.9 | 0.55 (0.51‐0.60) | 3.5 4.9 | 0.32 (0.28‐0.36) |
Admitted to ward, LAPS <20, did experience transfer to ICU or TCU | 4,341 (2.1) | 9.5 | 4.31 (3.90‐4.74) | 11.8 18.1 | 7.12 (6.61‐7.64) |
Transfers to a higher level of care at a different hospital, which in the KPMCP are usually planned, experienced lower mortality than transfers within the same hospital. For ward to TCU transfers, intra‐hospital transfers had a mortality of 12.1% while inter‐hospital transfers had a mortality of 5.7%. Corresponding rates for ward to ICU transfers were 21.7% and 11.2%, and for TCU to ICU transfers the rates were 25.9% and 12.5%, respectively.
Among patients initially admitted to the ward, a model to predict the occurrence of a transfer to a higher level of care (within 48 hours after admission) that included age, sex, admission type, primary condition, LAPS, COPS, and interaction terms had poor discrimination, with an area under the receiver operator characteristic (c statistic) of only 0.64. The c statistic for a model to predict transfer after 48 hours was 0.66. The corresponding models for TCU admits had c statistics of 0.67 and 0.68. All four models had poor calibration.
Discussion
Using automated bed history data permits characterizing a patient population with disproportionate mortality and LOS: intra‐hospital transfers to special care units (ICUs or TCUs). Indeed, the largest subset of these patients (those initially admitted to the ward or TCU) constituted only 3.7% of all admissions, but accounted for 24.2% of all ICU admissions, 21.7% of all hospital deaths, and 13.2% of all hospital days. These patients also had very elevated OEMRs and OMELOS. Models based on age, sex, preadmission laboratory test results, and comorbidities did not predict the occurrence of these transfers.
We performed multivariate analyses to explore the degree to which electronically assigned preadmission severity scores could predict these transfers. These analyses found that, compared to our ability to predict inpatient or 30‐day mortality at the time of admission, which is excellent, our ability to predict the occurrence of transfer after admission is much more limited. These results highlight the limitations of severity scores that rely on automated data, which may not have adequate discrimination when it comes to determining the risk of an adverse outcome within a narrow time frame. For example, among the 121,237 patients initially admitted to the ward who did not experience an intra‐hospital transfer, the mean LAPS was 18.9, while the mean LAPS among the 6,484 ward patients who did experience a transfer was 25.5. Differences between the mean and median LAPS, COPS, and predicted mortality risk among transferred and non‐transferred patients were significant (P < 0.0001 for all comparisons). However, examination of the distribution of LAPS, COPS, and predicted mortality risk between these two groups of patients showed considerable overlap.
Our methodology resembles Silber et al.'s27, 28 concept of failure to rescue in that it focuses on events occurring after hospitalization. Silber et al. argue that a hospital's quality can be measured by quantifying the degree to which patients who experience new problems are successfully rescued. Furthermore, quantification of those situations where rescue attempts are unsuccessful is felt to be superior to simply comparing raw or adjusted mortality rates because these are primarily determined by underlying case mix. The primary difference between Silber et al.'s approach and ours is at the level of detailthey specified a specific set of complications, whereas our measure is more generic and would include patients with many of the complications specified by Silber et al.27, 28
Most of the patients transferred to a higher level of care in our cohort survived (ie, were rescued), indicating that intensive care is beneficial. However, the fact that these patients had elevated OEMRs and OMELOS indicates that the real challenge facing hospitalists involves the timing of provision of a beneficial intervention. In theory, improved timing could result from earlier detection of problems, which is the underlying rationale for employing rapid response teams. However, the fact that our electronic tools (LAPS, COPS) cannot predict patient deteriorations within a narrow time frame suggests that early detection will remain a major challenge. Manually assigned vital signs scores designed for this purpose do not have good discrimination either.29, 30 This raises the possibility that, though patient groups may differ in terms of overall illness severity and mortality risk, differences at the individual patient level may be too subtle for clinicians to detect. Future research may thus need to focus on scores that combine laboratory data, vital signs, trends in data,31, 32 and newer proteomic markers (eg, procalcitonin).33 We also found that most transfers occurred early (within <72 hours), raising the possibility that at least some of these transfers may involve issues around triage rather than sudden deterioration.
Our study has important limitations. Due to resource constraints and limited data availability, we could not characterize the patients as well as might be desirable; in particular, we could not make full determinations of the actual reasons for patients' transfer for all patients. Broadly speaking, transfer to a higher level of care could be due to inappropriate triage, appropriate (preventive) transfer (which could include transfer to a more richly staffed unit for a specific procedure), relentless progression of disease despite maximal therapy, the occurrence of management errors, patient and family uncertainty about goals of care or inadequate understanding of treatment options and prognoses, or a combination of these factors. We could not make these distinctions with currently available electronic data. This is also true of postsurgical patients, in whom it is difficult to determine which transfers to intensive care might be planned (eg, in the case of surgical procedures where ICU care is anticipated) as opposed to the occurrence of a deterioration during or following surgery. Another major limitation of this study is our inability to identify code or no code status electronically. The elapsed LOS at time of death among patients who experienced a transfer to a higher level of care (as compared to patients who died in the ward without ever experiencing intra‐hospital transfer) suggests, but does not prove, that prolonged efforts were being made to keep them alive. We were also limited in terms of having access to other process data (eg, physician staffing levels, provision and timing of palliative care). Having ICU severity of illness scores would have permitted us to compare our cohort to those of other recent studies showing elevated mortality rates among transfer patients,911 but we have not yet developed that capability.
Consideration of our study findings suggests a possible research agenda that could be implemented by hospitalist researchers. This agenda should emphasize three areas: detection, intervention, and reflection.
With respect to detection, attention needs to be paid to better tools for quantifying patient risk at the time a decision to admit to the ward is made. It is likely that such tools will need to combine the attributes of our severity score (LAPS) with those of the manually assigned scores.30, 34 In some cases, use of these tools could lead a physician to change the locus of admission from the ward to the TCU or ICU, which could improve outcomes by ensuring more timely provision of intensive care. Since problems with initial triage could be due to factors other than the failure to suspect or anticipate impending instability, future research should also include a cognitive component (eg, quantifying what proportion of subsequent patient deteriorations could be ascribed to missed diagnoses35). Additional work also needs to be done on developing mathematical models that can inform electronic monitoring of ward (not just ICU) patients.
Research on interventions that hospitalists can use to prevent the need for intensive care or to improve the rescue rate should take two routes. The first is a disease‐specific route, which builds on the fact that a relatively small set of conditions (pneumonia, sepsis, gastrointestinal bleeding) account for most transfers to a higher level of care. Condition‐specific protocols, checklists, and bundles36 tailored to a ward environment (as opposed to the ICU or to the entire hospital) might prevent deteriorations in these patients, as has been reported for sepsis.37 The second route is to improve the overall capabilities of rapid response and code blue teams. Such research would need to include a more careful assessment of what commonalities exist among patients who were and were not successfully rescued by these teams. This approach would probably yield more insights than the current literature, which focuses on whether rapid response teams are a good thing or not.
Finally, research also needs to be performed on how hospitalists reflect on adverse outcomes among ward patients. Greater emphasis needs to be placed on moving beyond trigger tool approaches that rely on manual chart review. In an era of expanding use of electronic medical record systems, more work needs to be done on how to harness these to provide hospitalists with better quantitative and risk‐adjusted information. This information should not be limited to simply reporting rates of transfers and deaths. Rather, finer distinctions must be provided with respect of the type of patients (ie, more diagnostic detail), the clinical status of patients (ie, more physiologic detail), as well as the effects of including or excluding patients in whom therapeutic options may be limited (ie, do not resuscitate and comfort care patients) on reported rates. Ideally, researchers should develop better process and outcomes measures that could be tested in collaborative networks that include multiple nonacademic general medical‐surgical wards.
Acknowledgements
The authors thank Drs. Paul Feigenbaum, Alan Whippy, Joseph V. Selby, and Philip Madvig for reviewing the manuscript and Ms. Jennifer Calhoun for formatting the manuscript.
- To Err is Human: Building a Safer Health System.Washington, D. C.:National Academy Press;2000. , , .
- Institute for Healthcare Improvement. Protecting 5 million lives from harm. Available at: http://www.ihi.org/IHI/Programs/Campaign. Accessed June2010.
- Identifying poor‐quality hospitals. Can hospital mortality rates detect quality problems for medical diagnoses?Med Care.1996;34(8):737–753. , .
- Surgical mortality as an indicator of hospital quality: the problem with small sample size.JAMA.2004;292(7):847–851. , , .
- State of California Office of Statewide Health Planning and Development. AHRQ ‐ Inpatient quality indicators (IQIs) hospital inpatient mortality indicators for California. Available at: http://www.oshpd.ca.gov/HID/Products/PatDischargeData/AHRQ/iqi‐imi_overview.html. Accessed June2010.
- Enhancement of claims data to improve risk adjustment of hospital mortality.JAMA.2007;297(1):71–76. , , , et al.
- Variation in outcomes in Veterans Affairs intensive care units with a computerized severity measure.Crit Care Med.2005;33(5):930–939. , , , et al.
- Acute Physiology and Chronic Health Evaluation (APACHE) IV: hospital mortality assessment for today's critically ill patients.Crit Care Med.2006;34(5):1297–1310. , , , .
- Day of the week of intensive care admission and patient outcomes: a multisite regional evaluation.Med Care.2002;40(6):530–539. , , , .
- The hospital mortality of patients admitted to the ICU on weekends.Chest.2004;126(4):1292–1298. , , , et al.
- Mortality among patients admitted to intensive care units during weekday day shifts compared with “off” hours.Crit Care Med.2007;35(1):3–11. , , , et al.
- Septic shock: an analysis of outcomes for patients with onset on hospital wards versus intensive care units.Crit Care Med.1998;26(6):1020–1024. , , , et al.
- Clinical antecedents to in‐hospital cardiopulmonary arrest.Chest.1990;98(6):1388–1392. , , , , .
- Developing strategies to prevent inhospital cardiac arrest: analyzing responses of physicians and nurses in the hours before the event.Crit Care Med.1994;22(2):244–247. , .
- MERIT Study Investigators.Introduction of the medical emergency team (MET) system: a cluster‐randomized controlled trial.Lancet.2005;365(9477):2091–2097.
- Institute for Healthcare Improvement.The “MERIT” Trial of Medical Emergency Teams in Australia: An Analysis of Findings and Implications.Boston, MA:2005. Available on www.ihi.org
- Rapid response teams‐‐walk, don't run.JAMA.2006;296(13):1645–1647. , , .
- IHI Global Trigger Tool for Measuring Adverse Events.2nd ed.Cambridge, Massachusetts:Institute for Healthcare Improvement;2009. , .
- Risk adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases.Medical Care.2008;46(3):232–239. , , , , , .
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population.J Clin Epidemiol.2010;63(7):798–803. , , , .
- Linking automated databases for research in managed care settings.Ann Intern Med.1997;127(8 Pt 2):719–724. .
- Anticoagulation therapy for stroke prevention in atrial fibrillation: how well do randomized trials translate into clinical practice?JAMA.2003;290(20):2685–2692. , , , et al.
- Richardson score predicts short‐term adverse respiratory outcomes in newborns >/=34 weeks gestation.J Pediatr.2004;145(6):754–760. , , , et al.
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases.Am J Manag Care.2008;14(3):158–166. , , , et al.
- Statistical Analysis Software [computer program]. Version 8.Cary, NC:SAS Institute, Inc.;2000.
- Use and misuse of the receiver operating characteristic curve in risk prediction.Circulation.2007;115(7):928–935. .
- Hospital and patient characteristics associated with death after surgery. A study of adverse occurrence and failure to rescue.Med Care.1992;30(7):615–629. , , , .
- Comparing the contributions of groups of predictors: which outcomes vary with hospital rather than patient characteristics?J Am Stat Assoc.1995;90(429):7–18. , , .
- Beyond the intensive care unit: A review of interventions aimed at anticipating and preventing in‐hospital cardiopulmonary arrest.Resuscitation.2005;67(1):13–23. , .
- Reproducibility of physiological track‐and‐trigger warning systems for identifying at‐risk patients on the ward.Intensive Care Med.2007;33(4):619–624. , , .
- Serial evaluation of the SOFA score to predict outcome in critically ill patients.JAMA.2001;286(14):1754–1758. , , , , .
- Incorporation of Physiologic Trend and Interaction Effects in Neonatal Severity of Illness Scores: An Experiment Using a Variant of the Richardson Score.Intensive Care Med.2007;33(9):1602–1608. , , .
- Diagnostic and prognostic value of procalcitonin in patients with septic shock.Crit Care Med.2004;32(5):1166–1169. , , , et al.
- Identifying the sick: can biochemical measurements be used to aid decision making on presentation to the accident and emergency department.Br J Anaesth.2005;94(6):735–741. , , , et al.
- Improving patient care. The cognitive psychology of missed diagnoses.Ann Intern Med.2005;142(2):115–120. .
- Using care bundles to reduce in‐hospital mortality: quantitative survey.BMJ.2010;340:c1234. , , , , , .
- Effect of a rapid response system for patients in shock on time to treatment and mortality during 5 years.Crit Care Med.2007;35(11):2568–2575. , , , et al.
- To Err is Human: Building a Safer Health System.Washington, D. C.:National Academy Press;2000. , , .
- Institute for Healthcare Improvement. Protecting 5 million lives from harm. Available at: http://www.ihi.org/IHI/Programs/Campaign. Accessed June2010.
- Identifying poor‐quality hospitals. Can hospital mortality rates detect quality problems for medical diagnoses?Med Care.1996;34(8):737–753. , .
- Surgical mortality as an indicator of hospital quality: the problem with small sample size.JAMA.2004;292(7):847–851. , , .
- State of California Office of Statewide Health Planning and Development. AHRQ ‐ Inpatient quality indicators (IQIs) hospital inpatient mortality indicators for California. Available at: http://www.oshpd.ca.gov/HID/Products/PatDischargeData/AHRQ/iqi‐imi_overview.html. Accessed June2010.
- Enhancement of claims data to improve risk adjustment of hospital mortality.JAMA.2007;297(1):71–76. , , , et al.
- Variation in outcomes in Veterans Affairs intensive care units with a computerized severity measure.Crit Care Med.2005;33(5):930–939. , , , et al.
- Acute Physiology and Chronic Health Evaluation (APACHE) IV: hospital mortality assessment for today's critically ill patients.Crit Care Med.2006;34(5):1297–1310. , , , .
- Day of the week of intensive care admission and patient outcomes: a multisite regional evaluation.Med Care.2002;40(6):530–539. , , , .
- The hospital mortality of patients admitted to the ICU on weekends.Chest.2004;126(4):1292–1298. , , , et al.
- Mortality among patients admitted to intensive care units during weekday day shifts compared with “off” hours.Crit Care Med.2007;35(1):3–11. , , , et al.
- Septic shock: an analysis of outcomes for patients with onset on hospital wards versus intensive care units.Crit Care Med.1998;26(6):1020–1024. , , , et al.
- Clinical antecedents to in‐hospital cardiopulmonary arrest.Chest.1990;98(6):1388–1392. , , , , .
- Developing strategies to prevent inhospital cardiac arrest: analyzing responses of physicians and nurses in the hours before the event.Crit Care Med.1994;22(2):244–247. , .
- MERIT Study Investigators.Introduction of the medical emergency team (MET) system: a cluster‐randomized controlled trial.Lancet.2005;365(9477):2091–2097.
- Institute for Healthcare Improvement.The “MERIT” Trial of Medical Emergency Teams in Australia: An Analysis of Findings and Implications.Boston, MA:2005. Available on www.ihi.org
- Rapid response teams‐‐walk, don't run.JAMA.2006;296(13):1645–1647. , , .
- IHI Global Trigger Tool for Measuring Adverse Events.2nd ed.Cambridge, Massachusetts:Institute for Healthcare Improvement;2009. , .
- Risk adjusting hospital inpatient mortality using automated inpatient, outpatient, and laboratory databases.Medical Care.2008;46(3):232–239. , , , , , .
- The Kaiser Permanente inpatient risk adjustment methodology was valid in an external patient population.J Clin Epidemiol.2010;63(7):798–803. , , , .
- Linking automated databases for research in managed care settings.Ann Intern Med.1997;127(8 Pt 2):719–724. .
- Anticoagulation therapy for stroke prevention in atrial fibrillation: how well do randomized trials translate into clinical practice?JAMA.2003;290(20):2685–2692. , , , et al.
- Richardson score predicts short‐term adverse respiratory outcomes in newborns >/=34 weeks gestation.J Pediatr.2004;145(6):754–760. , , , et al.
- Risk adjusting community‐acquired pneumonia hospital outcomes using automated databases.Am J Manag Care.2008;14(3):158–166. , , , et al.
- Statistical Analysis Software [computer program]. Version 8.Cary, NC:SAS Institute, Inc.;2000.
- Use and misuse of the receiver operating characteristic curve in risk prediction.Circulation.2007;115(7):928–935. .
- Hospital and patient characteristics associated with death after surgery. A study of adverse occurrence and failure to rescue.Med Care.1992;30(7):615–629. , , , .
- Comparing the contributions of groups of predictors: which outcomes vary with hospital rather than patient characteristics?J Am Stat Assoc.1995;90(429):7–18. , , .
- Beyond the intensive care unit: A review of interventions aimed at anticipating and preventing in‐hospital cardiopulmonary arrest.Resuscitation.2005;67(1):13–23. , .
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