Abstract
Background and Purpose:
Observational data suggest that antiplatelet therapy after intracerebral hemorrhage (ICH) alleviates thromboembolic risk without increasing the risk of recurrent ICH. Given the paucity of data on the relationship between antiplatelet therapy after ICH and functional outcomes, we aimed to study this association in a multicenter cohort.
Methods:
We meta-analyzed data from: (1) the Massachusetts General Hospital (MGH) ICH registry (n=1,854), (2) the Virtual International Stroke Trials Archive (VISTA-ICH) database (n=762), and (3) the Yale stroke registry (n=185). Our exposure was antiplatelet therapy after ICH, which was modelled as a time varying covariate. Our primary outcomes were all-cause mortality and a composite of major disability or death (modified Rankin Scale score 4–6). We used Cox proportional regression analyses to estimate the hazard ratio (HR) of death or poor functional outcome as a function of antiplatelet therapy and random-effects meta-analysis to pool the estimated HRs across studies. Additional analyses stratified by hematoma location (lobar and deep ICH) were performed.
Results:
We included a total of 2,801 ICH patients, of whom 288 (10.3%) were started on antiplatelet medications after ICH. Median times to antiplatelet therapy ranged from 7 to 39 days. Antiplatelet therapy after ICH was not associated with mortality (HR, 0.85; 95% confidence interval [CI], 0.66–1.09), or death or major disability (HR, 0.83; 95% CI, 0.59–1.16) compared to patients not started on antiplatelet therapy. Similar results were obtained in additional analyses stratified by hematoma location.
Conclusions:
Antiplatelet therapy after ICH appeared safe, and was not associated with all-cause mortality or functional outcome, regardless of hematoma location. Randomized clinical trials are needed to determine the effects and harms of antiplatelet therapy after ICH.
Keywords: Intracerebral hemorrhage, Antiplatelet use, mortality, functional outcomes
Introduction
The relationship between prior antiplatelet therapy (APT) and intracerebral hemorrhage (ICH) outcomes has yet to be established, with studies yielding conflicting results.1, 2 Recent studies have failed to show an association between prior single APT and mortality or major disability in large ICH cohorts2, although dual APT may result in higher mortality.3 Furthermore, conflicting data on hematoma expansion4, 5 and concerns over ventriculostomy tract hemorrhage6 with APT have contributed to the hesitancy among clinicians to reinstate APT after ICH, even in the presence of strong indications.
Non-randomized data including our own suggest that restarting oral anticoagulation therapy after ICH is associated with a lower risk of ischemic stroke and improvement in functional outcomes.7, 8 Similarly, emerging observational data appear to favor the use of antiplatelet medications after ICH given the decreased incidence of thromboembolic events, without an apparent offsetting increase in the risk of ICH recurrence.9, 10 In fact, the Restart or Stop Antithrombotics (RESTART) randomized trial that evaluated APT resumption after ICH, showed that the risk of recurrent ICH was too small to exceed the established benefits of APT for secondary prevention.11 There is however, a paucity of evidence on the impact of APT on functional outcomes after ICH. In a recent analysis of patients enrolled in the Ethnic and Racial Variation in Intracerebral Hemorrhage (ERICH) study, restarting APT after ICH was initially associated with worse functional outcomes at 90 days.12 However, these findings were ultimately attributed to differences in baseline characteristics and possibly confounding by indication and no adverse association was observed in a propensity score matched analysis subsequently.12 In the absence of randomized trials, it is important to address the impact of APT on ICH disability. We therefore sought to leverage multiple ongoing observational studies of ICH to investigate whether APT after ICH was associated with mortality and functional outcome.
Materials and Methods
Study Design and Patients
We conducted a cohort study using data from 3 participating studies: 1) the single-center longitudinal study of ICH conducted at MGH in Boston, Massachusetts from 1994 onward and currently ongoing7; 2) the Virtual International Stroke Trials Archive-ICH (VISTA-ICH) established in 2010, which includes anonymized, individual patient-level data from completed ICH trials13; and 3) the single-center longitudinal ICH database of Yale University School of Medicine from 2013 and currently ongoing. Inclusion criteria in each study were: (1) diagnosis of acute primary (nontraumatic) ICH confirmed on computed tomography scan, (2) age 18 years or older at time of ICH, (3) no history of prior ICH, and 4) follow up data available at 90 days. We excluded patients with secondary causes of ICH such as trauma, vascular malformation, tumor or other mass lesions, cerebral venous sinus thrombosis, and hemorrhagic transformation of ischemic infarct. Additionally, patients on prior anticoagulant medications, ascertained on chart review, were also excluded from our analysis. Due to institutional data sharing restrictions, we analyzed the 3 cohorts separately. The study was approved by the institutional review boards (IRB) of Weill Cornell Medicine, Yale School of Medicine, and Massachusetts General Hospital (MGH).
Data Availability Statement
The VISTA-ICH dataset used in this analysis may be obtained after approval of a written proposal by the VISTA-ICH steering committee. The Yale stroke registry data may be shared with investigators after IRB approval. Given data sharing restrictions for the MGH data instituted by the IRB, this dataset cannot be shared with investigators. However, results of statistical analyses from the MGH ICH registry may be shared with other investigators.
Exposures and Outcomes
Our exposure was APT after ICH. Our primary outcomes were all-cause mortality and a composite of death or major disability, defined as modified Rankin Scale (mRS) scores of 4–6 assessed at 90 days. Follow-up information was obtained via a combination of telephone interviews, clinic visits, chart reviews, and social security death index database interrogations to identify unreported mortality events. All follow-up and outcome adjudications were conducted centrally by the coordinating center for each cohort.
Statistical Analysis
Overall Analysis Plan:
Due to institutional data sharing restrictions, we analyzed the 3 cohorts separately. We first performed Cox regression analyses in each cohort and subsequently meta-analyzed the hazard ratio (HR) obtained from the 3 studies to calculate a pooled HR for each outcome.
Analysis of the Individual Cohorts:
We defined APT as a time varying covariate. Specifically, they contributed person-time to the unexposed group (controls) from the time of the study entry until the start of APT, and then contributed to the exposed group from the time of stating APT to the end of follow-up. Our rationale for using this methodology was to mitigate the confounding effect of early deaths and variable timing of APT. Univariate and multivariate analyses were performed as detailed below within each individual study dataset. We used Cox regression models in the multivariate analysis assessing the relationship between APT and outcome (mortality and favorable outcome). To ensure uniformity in the statistical analyses across the unpublished cohorts, the Cox models were adjusted for pre-specified covariates that included age, sex, admission Glasgow Coma Scale score, admission hematoma volumes, presence of intraventricular hemorrhage, hematoma location, and prior APT, regardless of significance on the univariate analysis. We tested the proportional hazard assumption separately within each study by means of graphical checks and Schoenfeld residual based tests. Multicollinearity was assessed by computing variance inflation factors for all variables.
Meta-analysis Plan:
We performed a meta-analysis using results from the 3 cohorts to assess the association between APT and ICH outcomes using the pooled HR as the effect parameter for primary outcomes, and pooled OR for the secondary outcome. We used the adjusted HR (OR for mRS shift) and confidence intervals (CI) from each study to generate random-effects (DerSimonian-Laird) models to calculate the pooled HR or pooled OR, and generated forest plots.14, 15 The rationale of using the more conservative random-effects model was to account for the variability in effect sizes, design, and follow-up between the individual studies. We assessed heterogeneity using the Cochran Q test. Statistical analyses were performed using Stata (version 14.0, College Station, TX). All tests were two-tailed and p-values <0.05 (confidence intervals not including the null value of 1) were considered significant.
Additional Analyses:
We performed additional analyses to study the relationship between APT and primary outcomes, stratified by hematoma location (lobar and deep). Similar to the primary analysis, we first calculated the HR in the 3 cohorts for each outcome, and then meta-analyzed the HRs to generate a pooled HR.
Results
We included a total of 2,801 ICH patients, of whom 288 (10.3%) were started on APT after ICH. Demographics, comorbidities, and ICH characteristics of the 3 cohorts are shown in Table 1. There were no major significant differences in baseline demographics, hematoma characteristics, or clinical severity between the two groups. The median times from ICH diagnosis to use of APT were 11 days (interquartile range, [IQR], 7–15), 39 days (IQR, 11–66), and 7 days (IQR, 5–15), in the MGH, VISTA-ICH, and Yale cohorts, respectively. Among patients on APT after ICH, prior APT was reported in 73 (32.9%, MGH), 10 (23.8%, VISTA-ICH), and 1 (4.2%, Yale) patients (Table 2).
Table 1.
Patient Characteristics of the 3 Cohorts
| Characteristic | MGH | VISTA-ICH | Yale |
|---|---|---|---|
| Number of patients | 1854 | 762 | 185 |
| Demographics | |||
| Age, mean (SD), years | 71.3 (11.5) | 65.4 (12.3) | 67.2 (14.0) |
| Sex, male | 1023 (55.2) | 481 (63.1) | 93 (50.3) |
| Comorbidities | |||
| Hypertension | 1423 (76.8) | 634 (83.2) | 131 (70.8) |
| Diabetes mellitus | 411 (22.2) | 130 (17.1) | 42 (22.7) |
| Dyslipidemia | 646 (34.8) | 76 (9.9) | 60 (32.4) |
| Prior stroke | 392 (21.1) | 95 (12.3) | 21 (11.4) |
| Coronary artery disease | 353 (19.0) | 23 (3.0) | 16 (8.7) |
| Atrial fibrillation | 444 (23.9) | 31 (4.1) | 34 (18.4) |
| Peripheral vascular disease | 191 (10.3) | 17 (2.2) | 9 (4.9) |
| Prior antiplatelet use | 621 (33.5) | 103 (13.5) | 4 (2.2) |
| Index ICH characteristics | |||
| ICH volumea | 18.5 (9.3 – 31.2) | 14.9 (7.5–29.2) | 11.1 (3.9–26.7) |
| IVH present | 377 (20.3) | 229 (30.1) | 46 (27.1) |
| Glasgow Coma Scalea | 14 (13–15) | 15 (13–15) | 14 (10–15) |
| Hematoma location | |||
| Lobar | 849 (45.8) | 329 (43.2) | 73 (39.5) |
| Deep | 911 (49.1) | 398 (52.2) | 82 (44.3) |
| Infratentorial | 94 (5.1) | 35 (4.6) | 30 (16.2) |
| Follow-up data | |||
| APT after ICH | 222 (12.0) | 42 (5.5) | 24 (12.9) |
| Time to APT after ICH a | 11 (7–15) | 39 (11–66) | 7 (5–15) |
| Outcome data at 3 months | |||
| Mortality | 521 (28.1) | 152 (19.9) | 37 (20.7) |
| mRS 4–6 | 895 (48.3) | 313 (41.2) | 72 (38.9) |
Abbreviations: APT, Antiplatelet Therapy; ICH, Intracerebral Hemorrhage; IVH, Intraventricular Hemorrhage; MGH, Massachusetts General Hospital; mRS, modified Rankin Score; VISTA-ICH, Virtual International Stroke Trials Archive-ICH; SD, Standard Deviation.
All values displayed as count (%) unless otherwise specified
values displayed as median (interquartile range)
Table 2.
Patient Characteristics, Stratified by Antiplatelet Therapy after Intracerebral Hemorrhage
| Characteristic | MGH | VISTA-ICH | Yale | |||
|---|---|---|---|---|---|---|
| APT Use | No APT Use | APT Use | No APT Use | APT Use | No APT Use | |
| Number of patients | 222 | 1632 | 42 (5.5) | 720 (94.5) | 24 | 161 |
| Demographics | ||||||
| Age, mean (SD), years | 72.3 (10.3) | 70.5 (11.0) | 68.0 (9.2) | 64.5 (12.2) | 71.3 (10.5) | 66.6 (14.4) |
| Sex, male | 125 (56.3) | 898 (55.0) | 24 (57.1) | 457 (63.5) | 14 (58.3) | 79 (49.1) |
| Comorbidities | ||||||
| Hypertension | 167 (75.2) | 1256 (77.0) | 36 (85.7) | 598 (83.1) | 23 (95.8) | 108 (67.1) |
| Diabetes mellitus | 52 (23.4) | 359 (22.0) | 8 (19.1) | 122 (16.9) | 8 (33.3) | 34 (21.1) |
| Dyslipidemia | 78 (35.1) | 568 (34.8) | 2 (4.8) | 74 (10.3) | 10 (41.7) | 50 (31.1) |
| Prior stroke | 50 (22.5) | 342 (21.0) | 5 (11.9) | 90 (12.5) | 2 (8.3) | 19 (11.8) |
| Coronary artery disease | 49 (22.1) | 304 (18.6) | 2 (4.8) | 21 (2.9) | 4 (16.7) | 12 (7.5) |
| Atrial fibrillation | 63 (28.4) | 381 (23.3) | 3 (7.1) | 28 (3.9) | 11 (45.8) | 23 (14.3) |
| Peripheral vascular disease | 25 (11.3) | 166 (10.2) | 1 (2.4) | 16 (2.2) | 3 (12.5) | 6 (3.7) |
| Prior APT | 73 (32.9) | 548 (33.6) | 10 (23.8) | 93 (12.9) | 1 (4.2) | 3 (1.9) |
| Index ICH characteristics | ||||||
| ICH volumea | 17.3 (8.6 – 28.4) | 18.2 (9.1 – 30.2) | 10.9 (4.9–19.6) | 13.7 (6.8–25.9) | 8.1 (2.9–12.6) | 11.9 (3.8–28.7) |
| IVH present | 42 (18.9) | 335 (20.5) | 15 (35.7) | 214 (29.7) | 6 (27.3) | 40 (27.0) |
| Glasgow Coma Scalea | 14 (13–15) | 14 (13–15) | 15 (14–15) | 15 (13–15) | 15 (14–15) | 14 (8–15) |
| Hematoma location | ||||||
| Lobar | 95 (42.8) | 754 (46.2) | 16 (40.1) | 313 (43.5) | 8 (33.3) | 65 (39.8) |
| Deep | 116 (52.3) | 795 (48.7) | 24 (57.1) | 374 (51.9) | 10 (41.7) | 72 (44.8) |
| Infratentorial | 11 (5.0) | 83 (5.1) | 2 (7.4) | 33 (4.6) | 6 (25.0) | 24 (14.9) |
| Outcome data at 3 months | ||||||
| Mortality | 64 (28.9) | 457 (28.0) | 6 (14.3) | 146 (20.2) | 1 (4.4) | 36 (23.1) |
| mRS 4–6 | 98 (44.1) | 797 (48.8) | 18 (42.9) | 295 (40.9) | 11 (45.8) | 61 (37.9) |
Abbreviations: APT, Antiplatelet Therapy; ICH, Intracerebral Hemorrhage; IVH, Intraventricular Hemorrhage; MGH, Massachusetts General Hospital; mRS, modified Rankin Score; VISTA-ICH, Virtual International Stroke Trials Archive-ICH; SD, Standard Deviation.
All values displayed as count (%) unless otherwise specified
values displayed as median (interquartile range)
In the multivariable Cox models, we adjusted for age, sex, admission Glasgow Coma Scale score, admission hematoma volumes, presence of intraventricular hemorrhage, and prior APT. APT after ICH was not associated with mortality or the composite of death or major disability in any of the cohorts (Figure 1). In the meta-analysis of the 3 cohorts, APT after ICH was not associated with mortality compared to patients not started on APT (pooled HR, 0.85; 95% CI, 0.66–1.09) (Figure 1). There was no evidence of heterogeneity (I2 = 0, P value for Cochran Q = 0.61). Similarly, APT was not associated with death or major disability compared to patients not on APT (pooled HR, 0.83; 95% CI, 0.59–1.16). There was no evidence of heterogeneity (I2 = 0, P value for Cochran Q = 0.43).
Figure 1.
Forest plot of the association between antiplatelet therapy and mortality (top panel) and disability (bottom panel) after intracerebral hemorrhage. The meta-analysis was calculated using a random-effects model, with the pooled relative risk shown in the forest plot. Each square represents the point estimate of any given study’s effect size. The size of the squares is proportional to the inverse of the variance of the estimate, while the horizontal lines represent each study’s 95% confidence intervals. The diamond represents the pooled estimate with the width of the diamond representing the pooled 95% CI.
When stratified by hematoma location, APT use after ICH was not associated with mortality (HR, 0.85; 95% CI, 0.55–1.32) or mRS 4–6 (HR, 0.91; 95% CI, 0.64–1.31) in lobar ICH (Figure 2). Similarly, there was no association between APT and mortality (HR, 0.88; 95% CI, 0.65–1.20) or mRS 4–6 (HR, 0.89; 95% CI, 0.52–1.55) in deep ICH (Figure 3). We only included HRs from the MGH and VISTA-ICH cohorts for this analysis, since the Yale cohort lacked power to calculate a HR for lobar and deep ICH.
Figure 2.
Forest plot of the association between antiplatelet therapy and mortality (top panel) and disability (bottom panel) in lobar intracerebral hemorrhage.
Figure 3.
Forest plot of the association between antiplatelet therapy and mortality (top panel) and disability (bottom panel) in deep intracerebral hemorrhage.
In post hoc analyses, we studied ICH outcomes based on whether APT was resumed or started de novo. De novo starting of APT was not associated with ICH outcomes compared to patients not on APT (Figures I and II, online only data supplement). Similarly, APT resumption did not influence mortality or functional outcome compared to patients not on APT (Figures III and IV, online only data supplement). Furthermore, comparison of patients restarted on APT with those in whom APT was started de novo showed similar odds of outcome (Table I, online only data supplement).
Discussion
In this multi-center cohort study, APT after ICH appeared safe and was not associated with all-cause mortality or functional outcome, regardless of hematoma location. The observed median times to starting APT ranged from 7–39 days across the studies.
There is a paucity of data on the relationship between APT after ICH and functional outcomes. Our study corroborates the results of prior studies. In the ERICH cohort study, restarting APT was initially associated with lower rates of good functional outcome at 90 days.12 However, no such adverse association was observed after adjusting for differences in baseline characteristics using propensity matching, and APT did not influence functional outcomes or overall quality of life. A multicenter Italian study of patients with warfarin-related ICH also reported no difference in mortality among patients after reinstatement of APT, compared to those not on antithrombotic medications.16. One possible explanation for the lack of association between APT and disability is that more than half of the patients in our study had an mRS of 4–6 at 90 days, which reflects the severity of disability these patients had as their new baseline following the ICH. Given that the trajectory for recovery after ICH continues beyond the traditionally used cut off of 90 days and extends well into the first year17, 18, it is possible that a longer time frame for assessment of ICH outcomes could more accurately capture the true benefits of APT.
Population studies have shown that APT used in the setting of primary prevention is associated with an increased risk of intracranial hemorrhage compared to subjects not on APT, particularly in the elderly population.19 Clinicians therefore are concerned about the heightened risk of ICH recurrence, should APT be started after ICH. The fact that recent non-randomized studies have shown that reinstitution of oral anticoagulation therapy, a class of medications with a higher hemorrhagic risk than APT, was not associated with a heightened risk of recurrent ICH offers a countervailing observation.7, 8, 16 Furthermore, in a meta-analysis of observational studies, reinstatement of APT among ICH patients with atrial fibrillation did not increase the risk of recurrent ICH compared to patients not on APT.10 One possible explanation for this apparent conflict is the possibility that secondary blood pressure control is more robust post ICH than in the care rendered for primary prevention of stroke.
We also observed a variation in the timing of APT after ICH where the median time to APT ranged from 7–39 days. This reflects the current evidence where the American Heart Association guidelines recommend consideration of antiplatelet monotherapy after ICH regardless of hematoma location, but do not specify the optimal timeframe for APT resumption.20 In fact, this discrepancy in the timing of APT after ICH is also evident from prior studies. For instance, in the RESTART trial, the median time to resumption of APT was 76 days (IQR, 29–146)21, while in an observational study from Scotland reported the median time from ICH discharge to APT use was nearly 15 months22. Similarly, a retrospective study from the Swedish Stroke Registry showed that rates of APT prescription were 43.6% and 17.5% within 1 year after ICH among patients with and without atrial fibrillation, respectively.23,24
Our study has a few limitations. First, our study is subject to biases from the non-randomized and retrospective design. Despite rigorous adjustment of baseline severity and comorbidities, our study is subject to confounding by indication in that clinicians were more likely to start APT in patients perceived to have modest to good outcome such as younger patients, and those with smaller hematoma volumes. Moreover, specific indications that prompted use of APT were also not available which likely also introduced bias in the study. We tried to partially address this issue including these factors as covariates in the Cox regression models. Second, to account for variation in the timing of APT, we used APT as a time varying covariate, to minimize immortal time bias. However, medication compliance could not be reliably assessed in our study and may have hampered our ability to accurately capture time on APT. Third, we did not have information on the specific cause of death, thrombotic events, or ICH recurrence. Finally, information on presence of cerebral microbleeds or amyloid angiopathy, which may have influenced decisions on starting APT, was not available in all cohorts.
In summary, APT after ICH was not associated with mortality or severe disability in our study. Furthermore, APT use appeared to be safe in both lobar and deep ICH. Large randomized controlled trials are warranted to better elucidate long-term beneficial and harmful effects of APT after ICH.
Supplementary Material
Acknowledgements:
Funding/Support: SBM is supported by the NIH (K23NS105948) and the Leon Levy Foundation. AB is supported by the NIH (K23NS100816). GJF is supported by the NIH (K76AG059992, R03NS112859), the American Heart Association (18IDDG34280056), the Yale Pepper Scholar Award (P30AG021342) and the Neurocritical Care Society Research Fellowship. LS is supported by the NIH (R01NS095993, R01NS097728). BBN is supported by the NIH (K23NS091395) and the Florence Gould Endowment for Discovery in Stroke. HK is supported by the NIH (U01NS095869 and R01NS097443) and the Michael Goldberg Research Fund. DFH is supported by the NIH (U01NS080824 and U24TR001609). WCZ is supported by the NIH (1U01NS08082). JR is supported by the NIH (R01NS036695, UM1HG008895, R01NS093870, and R24NS092983), and OneMind. KNS is supported by the NIH (U24NS107215, U24NS107136, RO1NR018335, and U01NS106513), Novartis, and Bard.
LHS reports past consulting fees and non-financial support for Genentech unrelated to the subject of this work. DFH reports personal fees from Op2Lysis, personal fees from BrainScope, personal fees from Neurotrope, and non-financial support from Genentech outside the submitted work. WCZ receives consulting fees from C.R. Bard, Inc. outside of the area of work commented on here. JR reports consulting fees from Boehringer Ingelheim, Pfizer, New Beta Innovation. KNS reports grants from Hyperfine, Biogen, and Astrocyte unrelated to this work.
Appendix:
VISTA-ICH Steering Committee Collaborators: D.F. Hanley (Chair), K. Butcher, S. Davis, B. Gregson, K.R. Lees, P. Lyden, S. Mayer, K. Muir, and T. Steiner.
Footnotes
Disclosures: All other authors report no conflict of interest for this study.
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Associated Data
This section collects any data citations, data availability statements, or supplementary materials included in this article.
Supplementary Materials
Data Availability Statement
The VISTA-ICH dataset used in this analysis may be obtained after approval of a written proposal by the VISTA-ICH steering committee. The Yale stroke registry data may be shared with investigators after IRB approval. Given data sharing restrictions for the MGH data instituted by the IRB, this dataset cannot be shared with investigators. However, results of statistical analyses from the MGH ICH registry may be shared with other investigators.