Keywords
pulmonary embolism - stroke - prevention - thrombosis - venous thromboembolism
Introduction
Postacute care of anticoagulant-associated major hemorrhage is a challenging clinical
scenario as there is significant uncertainty about whether and when to restart anticoagulants.
Recently, researchers have focused on early anticoagulation restart after anticoagulant-associated
major hemorrhage as a strategy to reduce thrombotic events.[1]
[2] Several retrospective analyses, registry studies, and meta-analyses support restarting,
mostly in atrial fibrillation populations, to reduce thrombotic events, though with
some increase in rebleeding.[3]
[4]
[5]
[6]
[7]
[8] The ANNEXA-4 study of patients with factor Xa (FXa) inhibitor-associated severe
and potentially life-threatening bleeding receiving andexanet alfa was performed at
63 medical centers in North America and Europe, leading to regulatory approval.[9] ANNEXA-4 data provide an opportunity to study prospectively gathered data on 352
patients with FXa inhibitor-associated major bleeding, many of whom were restarted
on oral anticoagulation at the discretion of their treating physicians.
We previously reported in the ANNEXA-4 trial that no thrombotic events occurred after
restart of therapeutic oral anticoagulation,[9] but did not examine the association of restarting with bleeding or death or with
any composites that attempt to detect a net clinical benefit. The ANNEXA-4 dataset
offers granular details on anticoagulant restarting behavior, thrombotic and bleeding
events, and death through 30 days with very few patient dropouts. This allows more
thorough analysis than retrospective hospital datasets which often lose track of patients
at discharge, and thus fail to capture subsequent thrombotic and bleeding events.
The present study is a posthoc analysis evaluating the association between the occurrence
of thrombotic events, rebleeding, and death with the restarting of oral anticoagulation
in patients with FXa inhibitor-associated major bleeding.
Methods
The ANNEXA-4 study and patient population have been previously described.[9] Briefly, patients presented with FXa inhibitor-associated major bleeding, of which,
227 (64%) had intracranial hemorrhage (ICH; 128 spontaneous, 99 traumatic), 90 (26%)
had gastrointestinal (GI) bleeding, and 35 (10%) had other major bleeding ([Table 1]). Baseline characteristics, such as demographics, bleeding site, anti-FXa activity,
indication for anticoagulation, CHA2DS2-VASc score, history of thromboembolism and bleeding, and time from last dose to bleeding
onset, were assessed.
Table 1
Baseline characteristics for anticoagulation restart and non-restart (red for those
controlled for multivariate analysis)
|
Variable
|
Total (N = 352)
|
OAC non-restart (n = 252)
|
OAC restart (n = 100)
|
p-Value
|
|
Age, y
|
|
Median (IQR)
|
79.0 (72.0–84.0)
|
80.0 (73.0–86.0)
|
76.5 (71.0–82.0)
|
0.005
|
|
Mean (SD)
|
77.36 (10.80)
|
78.38 (10.41)
|
74.79 (11.36)
|
|
|
Race, No. (%)
|
|
|
|
0.171
|
|
White
|
307 (87.2)
|
223 (88.5)
|
84 (84.0)
|
|
|
Black/African-American
|
29 (8.2)
|
17 (6.7)
|
12 (12.0)
|
|
|
Other
|
7 (2.0)
|
4 (1.6)
|
3 (3.0)
|
|
|
Ethnicity, No. (%)
|
|
|
|
|
|
Non-Hispanic
|
327 (92.9)
|
231 (91.7)
|
96 (96.0)
|
0.125
|
|
Sex, No. (%)
|
|
|
|
|
|
Male
|
187 (53.1)
|
130 (51.6)
|
57 (57.0)
|
0.359
|
|
Primary bleed site, No. (%)
|
|
|
|
|
|
ICrH
|
227 (64.5)
|
186 (73.8)
|
41 (41.0)
|
<0.001
|
|
GI
|
90 (25.6)
|
50 (19.8)
|
40 (40.0)
|
<0.001
|
|
Other
|
35 (9.9)
|
16 (6.3)
|
19 (19.0)
|
<0.001
|
|
Indication for FXaI, No. (%)
|
|
|
|
|
|
AF
|
280 (79.5)
|
204 (81.0)
|
76 (76.0)
|
0.299
|
|
VTE
|
71 (20.2)
|
47 (18.7)
|
24 (24.0)
|
0.259
|
|
Other
|
26 (7.4)
|
20 (7.9)
|
6 (6.0)
|
0.531
|
|
Positive history of bleeding in the past 6 months, No. (%)
|
25 (7.1)
|
17 (6.7)
|
8 (8.0)
|
0.701
|
|
History of any thrombotic events in the past 6 months, No. (%)
|
|
MI
|
8 (2.3)
|
5 (2.0)
|
3 (3.0)
|
0.693
|
|
Stroke
|
305 (86.6)
|
217 (86.1)
|
88 (88.0)
|
0.762
|
|
TIA
|
332 (94.3)
|
235 (93.3)
|
97 (97.0)
|
0.168
|
|
Angina
|
24 (6.8)
|
15 (6.0)
|
9 (9.0)
|
0.306
|
|
VTE
|
38 (10.8)
|
31 (12.3)
|
7 (7.0)
|
0.156
|
|
Antiplatelet medications, No. (%)
|
92 (26.1)
|
64 (25.4)
|
28 (28.0)
|
0.616
|
|
Bleeding is a result of trauma, No. (%)
|
113 (32.1)
|
82 (32.5)
|
31 (31.0)
|
0.780
|
|
Time from the last FXaI dose to the onset of bleeding, n
[a]
|
163
|
110
|
53
|
|
|
>8 h, No. (%)
|
43 (26.4)
|
30 (27.3)
|
13 (24.5)
|
0.710
|
|
CHA2DS2-VASc score, n
|
352
|
252
|
100
|
|
|
Median (IQR)
|
4.0 (3.0–6.0)
|
4.0 (3.0–6.0)
|
4.0 (3.0–5.0)
|
0.198
|
Abbreviations: AF, atrial fibrillation; FXaI, factor Xa inhibitor; GI, gastrointestinal;
ICrH, intracranial hemorrhage; IQR, interquartile range; MI, myocardial infarction;
OAC, oral anticoagulation; TIA, transient ischemic attack; VTE, venous thromboembolism.
a Percentages are based on number of patients with available data (n) in the corresponding variable as the denominator (the denominator is at the top
of the column).
The follow-up started on the day of hemorrhage and ended 30 days later. For this analysis,
restart of anticoagulation was defined as administration of therapeutic dose oral
anticoagulants. The decision to restart anticoagulation was made by the treating physician
based on clinical judgment for each patient and could occur during the hospital stay
or at any time to 30-day end of study. We did not examine use of other forms of anticoagulation,
such as prophylactic dose low-molecular-weight heparin administered for venous thromboembolism
(VTE) prevention during hospitalization. Four outcomes were assessed: thrombotic events,
rebleeding, a composite of thrombotic events or rebleeding, and a composite of thrombotic
events, rebleeding, or death, which was designed to represent a net clinical benefit.
Thrombotic events were primary safety outcomes in ANNEXA-4, i.e., VTE, ischemic stroke,
other systemic embolism, and myocardial infarction. Symptomatic thrombotic events
were reported by site investigators and adjudicated by an independent academic committee.
Number and types of thrombotic events have been previously published.[9]
Rebleeding events were defined as bleeding at any site severe enough to be categorized
by the investigator as a serious adverse event (SAE; e.g., death, a life-threatening,
inpatient hospitalization, prolongation of existing hospitalization). Bleeding events
were evaluated posthoc for whether they represented true bleeding complications by
two independent reviewers (T.J.M. and J.W.E.), and any disagreements were adjudicated
by a third reviewer (M.C.). This was done by reviewing all SAEs related to bleeding
and applying clinical judgment as to whether they represented a bleeding event. Although
reviewers could not remain entirely agnostic to whether the patient had been restarted
on anticoagulation as it was sometimes mentioned in the SAE report, they were asked
only to determine whether a bleeding event had occurred, not to establish relatedness.
Baseline characteristics were summarized by patient groups defined by whether anticoagulation
was restarted or not. Categorical variables were reported as counts (percentages)
and compared between the two groups by Pearson's chi-squared test. Continuous variables
were summarized as mean (standard deviation) and/or median (interquartile range [IQR])
and compared between groups using the Wilcoxon rank-sum test. Statistical significance
was determined at the two-sided α = 0.05. Variables identified as significantly different
between the two groups ([Table 1]) were included in all multivariate Cox models.
As anticoagulation restart could occur any time during follow-up, the exposure variable
is time-dependent. As a first step, we conducted overall (naïve) tests of the relationship
between restart and various outcomes irrespective of the time of restart. In this
univariate analysis, we used a two-sided, log-rank test to compare the two treatment
groups for the survival distributions of the events. Cox proportional hazards regression
with the restart indicator as the only predictor was performed to estimate the hazard
ratio (HR). The 95% confidence interval (CI) was constructed by inverting the partial-likelihood
score test under the Cox model.[10] For the test of thrombotic events, in which a quasi-complete separation feature
presented (i.e., no thrombotic events after restart), Peto's method[11] was used to estimate HR and CI based on the log-rank test result. Multivariate Cox
proportional hazards regression analysis was also performed to adjust for possible
confounders. In the Cox regression models, parameter estimations were obtained using
a traditional maximum partial likelihood method. When quasi-complete separation occurred
for any analysis where no events occurred in one exposure group (such as after restart),
the Firth penalized partial likelihood method[12]
[13] for rare events was used to correct the bias. Statistical analyses were performed
using SAS software, version 9.4, and R Statistical Software.[14]
To account for the time-dependent nature of the restart treatment and to eliminate
the guarantee-time bias that it might bring,[15] we then applied landmark analysis and time-dependent covariate Cox regression methods
to examine the association between anticoagulation restart timing and outcomes.
Landmark Analysis
Landmark analysis is a commonly used approach to avoid the guarantee-time bias induced
by the time-dependent nature of the treatment.[16] This approach establishes a landmark time point during the follow-up period and
examines patients who are event free (no bleeding after day 3 and no thrombotic event
or death) prior to the landmark for a relationship between exposures occurring prior
to landmark (such as restarting) and outcomes occurring afterward (such as thrombotic
or bleeding events or death).[17] In our analysis, landmark time points were set a priori at day 3, 5, and 14 within
the 30-day follow-up period. The earliest landmark was chosen based on prior work
suggesting most rebleeding events occur within 3 days of index bleeding,[1] but few patients were restarted by day 3. We anticipated in our statistical analysis
plan that this would limit our ability to detect differences in outcomes. The day
5 landmark was an approximation slightly later to allow for more restarted patients.
The day 14 landmark was chosen because it roughly bisected the follow-up period. We
suspected this would give it the most power to detect differences. Later landmarks
would reduce power as they reduce the period of study between the landmark and the
30-day end of surveillance and thus the number of outcome events. Later landmarks
also exclude more patients, i.e., those with events prior to the landmark, reducing
power.
For each landmark analysis, patients were divided into restarted and non-restarted
groups based on whether they had restarted therapeutic oral anticoagulation prior
to the landmark time. Kaplan–Meier survival curves stratified by whether oral anticoagulation
was restarted were generated. A two-sided, log-rank test was applied to compare the
survival distribution of the events. Similar to the overall (naïve) analysis, we applied
both univariate (the restart indicator as the only predictor) and multivariate Cox
proportional hazards regression to obtain the unadjusted and adjusted estimations
of the HRs, p-values, and 95% CIs.
Time-Dependent Cox Regression
Time-dependent Cox hazard models restructured the exposure data to strictly account
for time spent restarted and not restarted on anticoagulation. In this model, exposure
to anticoagulation restart was regarded as a time-dependent variable. It was initially
assigned to be zero and irreversibly attained the value of one as soon as the patient
received therapeutic oral anticoagulation treatment. We applied the multivariate Cox
proportional hazards model and adjusted for the identified potential confounders to
investigate the relative risk of the various events between treatment groups.
We ignored the bleeding events in the first 3 days in both models because nearly all
were complications of the index bleeding events, and they confounded the model (details
below).
Subgroups
We performed the same analyses on the ICH subgroup (n = 227).
Results
Restarting: Treating physicians restarted oral anticoagulation in 100 of 352 (28%) patients
in the 30-day follow-up period. Of the 100 patients who were restarted, the median
restart time was 10 days (IQR: 5–16 days; [Supplementary Table S1] [available in the online version]) after the index bleeding event. Twelve patients
restarted within 3 days; 30 within 5 days; and 67 within 14 days. Patients were primarily
restarted on direct oral anticoagulants: 48% apixaban, 29% rivaroxaban, 4% dabigatran,
and 2% edoxaban. The remaining 17% were restarted on vitamin K antagonists (see [Supplementary Table S2] [available in the online version] for doses/percentages, [Supplementary Table S3] [available in the online version] for a full list and day of restart, and [Supplementary Table S4] [available in the online version] for baseline break down of outcomes by restarted
or not groups).
Patients who restarted were younger (median: 76.5 vs. 80.0 years; p = 0.005) and less likely to have ICH (41/100; 41% of restarted had ICH vs. 186/252;
73.8% of not restarted had ICH; p < 0.001). No other baseline characteristics were significantly associated with the
decision to restart using a p-value cut-off of 0.10. Thus, we included both age and site of index bleeding as covariates
in all multivariate Cox proportional hazards models.
Thrombotic events: Of 352 patients, 34 (9.7%) developed at least one thrombotic event in the 30-day
surveillance period. Of these 34 patients, the median time to first thrombotic event
was 10 days (IQR: 2–18 days; [Supplementary Table S1] [available in the online version]). Ischemic strokes (13 cases, 3.7%) and deep vein
thrombosis (13 cases, 3.7%) were the most frequent thrombotic event types, while myocardial
infarctions (7 cases, 2.0%), pulmonary embolism (5 cases, 1.4%), and transient ischemic
attacks (1 case, 0.3%) were less common ([Supplementary Table S1] [available in the online version]). There were 40 total thrombotic events in 34
patients. The thrombotic event rates in patients with ICH, GI bleeding, and other
bleeding were 9.3% (21 of 227), 6.7% (6 of 90), and 20% (7 of 35), respectively. No
thrombotic events occurred after restart of therapeutic anticoagulation; however,
eight thrombotic events (8% of those eventually restarted; 2.3%, 8 of 352 of the entire
sample) occurred in patients prior to restart ([Supplementary Table S1] [available in the online version]).
Bleeding events: The three reviewers (T.J.M., J.W.E., and M.C.) deemed that 29 (8.2%) patients had
a serious bleeding event (median time to first event: 3 days; IQR: 1–8 days; [Supplementary Table S1] [available in the online version]) (there were no disagreements in the adjudications).
Of the 14 bleeds in the first 3 days, 12 were either extensions of the index bleed
or new areas of bleeding discovered on computed tomography within the first 48 hours.
Of the other two, one developed melena on day 3, and the other had post-GI bleeding
anemia discovered on day 2. (None of these 14 patients had been restarted on oral
anticoagulation at the time of bleeding. Bleeding events occurring within 3 days were
not included in the analysis, but the patients who had those bleeding events within
the first 3 days were included in the subsequent analysis, i.e., they could still
go on to be restarted or not and have bleeding and thrombotic events or death as part
of both models.) The remaining 15 bleeding events (4.3% of 347 still living after
day 3) were incorporated in the models.
[Fig. 1] illustrates the cumulative thrombotic and bleeding events and a composite of the
two in the full cohort.
Fig. 1 Total cumulative incidence rates of thrombotic and rebleeding events (both restarted
and not).
Death: In the 30-day follow-up period, 49 patients (14%) died. The median time to death
was 12 days (IQR: 8–18; [Supplementary Table S1] [available in the online version]).
Landmark Analysis
The 14-day landmark analysis included 67 restarted and 234 non-restarted patients
([Table 2]). In the restarted group, none had thrombotic events, three patients had a bleeding
event and three patients died. In the non-restarted group, 12 patients had thrombotic
events, two patients had a bleeding event, and 17 patients died. In the unadjusted
analysis, restart before the 14-day landmark was not significantly associated with
reduced thrombotic events (p = 0.06) ([Fig. 2] and [Supplementary Table S5] [available in the online version]). After adjustment for age and bleeding site,
analyses at the 14-day landmark found that restart was associated with reduced thrombotic
events (HR: 0.112; 95% CI: 0.001–0.944; p = 0.043) and increased bleeding (HR: 8.39; 95% CI: 1.13–62.29; p = 0.037; [Table 2]). The composite outcomes and death alone were not significant at 14 days, and in
the other analyses done at the day 3 and day 5 landmarks; there were no significant
associations between restarting and the study outcomes, including death alone.
Table 2
Landmark analysis on day 14; multivariate analysis of anticoagulation restart versus
non-restart[a] (significant findings in red)
|
Variable
|
Coefficient
|
Standard error
|
p-Value
|
HR (95% CI)[b]
|
|
Rebleeding: for restart (n = 67; # of events: 3 [4.5%]); for non-restart (n = 234; # of events: 2 [0.9%])
|
|
Indicator of full OAC restart within 14 days (yes = 1, no = 0)
|
2.128
|
1.023
|
0.037
|
8.394 (1.131–62.288)
|
|
Age at screen
|
0.070
|
0.062
|
0.263
|
1.072 (0.949–1.211)
|
|
Initial primary bleeding site: ICrH
|
0.424
|
1.009
|
0.674
|
1.528 (0.211–11.049)
|
|
Thrombotic event[a]: for restart (n = 67; # of events: 0 [0.0%]); for non-restart (n = 234; # of events: 12 [5.1%])
|
|
Indicator of full OAC restart within 14 days (yes = 1, no = 0)
|
–2.186
|
1.531
|
0.043
|
0.112 (0.001–0.944)
|
|
Age at screen
|
0.019
|
0.030
|
0.526
|
1.019 (0.966–1.088)
|
|
Initial primary bleeding site: ICrH
|
–0.542
|
0.622
|
0.378
|
0.582 (0.188–2.073)
|
|
Death: for restart (n = 67; # of events: 3 [4.5%]); for non-restart (n = 234; # of events: 17 [7.3%])
|
|
Indicator of full OAC restart within 14 days (yes = 1, no = 0)
|
–0.230
|
0.683
|
0.737
|
0.795 (0.209–3.029)
|
|
Age at screen
|
0.104
|
0.032
|
0.001
|
1.110 (1.041–1.183)
|
|
Initial primary bleeding site: ICrH
|
–0.152
|
0.530
|
0.775
|
0.859 (0.304–2.430)
|
|
Composite of rebleeding and thrombotic event: for restart (N = 67; # of events: 3 [4.5%]); for non-restart (N = 234; # of events: 14 [6.0%])
|
|
Indicator of full OAC restart within 14 days (yes = 1, no = 0)
|
–0.262
|
0.687
|
0.703
|
0.770 (0.200–2.958)
|
|
Age at screen
|
0.033
|
0.028
|
0.226
|
1.034 (0.979–1.092)
|
|
Initial primary bleeding site: ICrH
|
–0.227
|
0.554
|
0.682
|
0.797 (0.269–2.360)
|
|
Composite of rebleeding, thrombotic event, and death: for restart (n = 67; # of events: 6 [9.0%]); for non-restart (n = 234; # of events: 29 [12.4%])
|
|
Indicator of full OAC restart within 14 days (yes = 1, no = 0)
|
–0.143
|
0.489
|
0.770
|
0.867 (0.332–2.261)
|
|
Age at screen
|
0.060
|
0.022
|
0.006
|
1.062 (1.018–1.108)
|
|
Initial primary bleeding site: ICrH
|
–0.108
|
0.397
|
0.786
|
0.898 (0.412–1.956)
|
Abbreviations: CI, confidence interval; HR, hazard ratio; ICrH, intracranial hemorrhage;
OAC, oral anticoagulation.
a Model was fitted by Proc PHREG (SAS9.4) using the proportional hazard Cox model except
for outcome thrombotic event.
b Firth correction using coxphf() in coxphf package of R was used for outcome thrombotic
event to correct bias due to quasi-complete separation feature in the data.
Fig. 2 Kaplan–Meier curve of thrombotic events for landmark analysis on day 14.
Cox Models Using Time-Dependent Exposure to Restart
In the Cox model, restarting was associated with reduced thrombotic events (HR = 0.071,
95% CI: 0.001–0.527; p = 0.004), and also with a reduction in the triple composite including death (HR = 0.384;
95% CI: 0.161–0.915; p = 0.031) ([Table 3]). There was no association of restart with bleeding or death in this model.
Table 3
Multivariate analysis of time-dependent Cox proportional hazards model of restart[a] (significant findings in red)
|
Variable
|
Coefficient
|
Standard error
|
p-Value
|
HR (95% CI)[b]
|
|
Multivariate analysis of outcome rebleeding
|
|
Indicator of restart OAC
|
0.512
|
0.721
|
0.478
|
1.668 (0.406–6.857)
|
|
Age at screen
|
–0.002
|
0.025
|
0.931
|
0.998 (0.951–1.047)
|
|
Initial primary bleeding site: ICrH
|
0.242
|
0.596
|
0.685
|
1.274 (0.396–4.093)
|
|
Multivariate analysis of outcome thrombotic event[b]
|
|
Indicator of restart OAC
|
–2.651
|
1.475
|
0.004
|
0.071 (0.001–0.527)
|
|
Age at screen
|
0.011
|
0.017
|
0.500
|
1.011 (0.980–1.048)
|
|
Initial primary bleeding site: ICrH
|
–0.449
|
0.361
|
0.216
|
0.638 (0.322–1.314)
|
|
Multivariate analysis of outcome death
|
|
Indicator of restart OAC
|
–1.040
|
0.619
|
0.093
|
0.354 (0.105–1.191)
|
|
Age at screen
|
0.066
|
0.018
|
<0.001
|
1.068 (1.031–1.107)
|
|
Initial primary bleeding site: ICrH
|
–0.162
|
0.320
|
0.613
|
0.850 (0.454–1.594)
|
|
Multivariate analysis of composite outcome of rebleeding and thrombotic events
|
|
Indicator of restart OAC
|
–1.032
|
0.623
|
0.098
|
0.356 (0.105–1.208)
|
|
Age at screen
|
0.011
|
0.015
|
0.445
|
1.011 (0.983–1.040)
|
|
Initial primary bleeding site: ICrH
|
–0.300
|
0.313
|
0.337
|
0.740 (0.401–1.367)
|
|
Multivariate analysis of composite outcome of rebleeding, thrombotic events, and death
|
|
Indicator of restart OAC
|
–0.958
|
0.443
|
0.031
|
0.384 (0.161–0.915)
|
|
Age at screen
|
0.031
|
0.012
|
0.011
|
1.031 (1.007–1.056)
|
|
Initial primary bleeding site: ICrH
|
–0.227
|
0.239
|
0.343
|
0.797 (0.499–1.274)
|
Abbreviations: CI, confidence interval; HR, hazard ratio; ICrH, intracranial hemorrhage;
OAC, oral anticoagulation.
a Model was fitted by Proc PHREG (SAS9.4) using the proportional hazard Cox model except
for outcome thrombotic event.
b Firth correction using coxphf() in coxphf package of R was used for outcome thrombotic
event to correct bias due to quasi-complete separation feature in the data.
Subgroups
Neither the landmarks nor Cox time-dependent models had any significant findings in
the ICH subgroup.
Discussion
Our data suggest that restarting may be associated with fewer thrombotic events and
increased bleeding. The time-dependent Cox model triple composite suggests that the
overall trade-off with restart when including death may be beneficial, but this needs
to be tested in prospective randomized trials.
There is significant uncertainty about whether and when to restart anticoagulants
after major hemorrhage. Thrombotic risk is high in the setting of the bleed itself,
particularly in ICH.[18] Added to this are patients' high baseline risks (i.e., the reason they were on an
anticoagulant before the index bleeding) and the risk of immobility after the bleeding
episode. This risk is cumulative, rising over time.[19] Bleeding risk, particularly hematoma expansion in ICH, is high in the first 72 hours
and less common thereafter.[1] Deciding whether and when to restart anticoagulation must balance those opposing
risks in any patient.
Some evidence suggests that restarting anticoagulation after major hemorrhage improves
outcomes, reducing thrombotic events and mortality, though with an increase in recurrent
bleeding.[3]
[4]
[5]
[6]
[7]
[8] Observational data to assess therapeutic effects are subject to confounding by indication
(e.g., lower risk patients are restarted earlier and more often, leading to better
outcomes), as we are in these models. However, such data are important in hypothesis
generation. Currently, there are no published randomized clinical trials, though several
addressing the question of whether to restart, when to restart or both, with bleeding,
thrombosis, and/or composite outcomes, are ongoing or about to begin (SoSTART [NCT03153150],
ENRICH-AF [NCT03950076], RESTART tICrH [NCT04229758], and ASPIRE [NCT03907046]). We
use these composite outcomes in our analysis as they closely approximate the clinical
decision that must be made for restarting anticoagulation, a balance between bleeding
and thrombosis, so they are clinically meaningful. Composite outcomes also increase
power as more events allow for the possibility of larger, more readily detectable
differences between groups.
All the adjudicated events in this dataset were serious but did vary in degree of
severity. This does cut both ways with bleeding and thrombosis events, i.e., an ICH
expansion may be worse than a deep vein thrombosis, but an ischemic stroke is generally
worse than a GI bleed. Balancing the severity of bleeding and thrombotic components
of composite outcomes has never been done even in prospective trials (though it is
a secondary analysis in Restart TICrH[20]) and would not be possible in this model. It requires correlating the events with
functional scales such as modified Rankin Scales performed later and does raise the
problem of what scales to use across the several outcome events, stroke, myocardial
infarction, pulmonary embolism, deep vein thrombosis, GI hemorrhage, recurrent ICH,
and other hemorrhage. No single functional scale is validated for all of them, and
using specific scales for each makes cross-comparison infeasible. One could simply
power on functional scales instead of events, but this requires samples in the thousands,
which are simply not feasible in a relatively infrequent disease. (For example, ANNEXA-4
required a global network of 63 sites more than 3 years to accrue 352 anticoagulated
major bleeding patients.[9]) A sample powered on mortality would have to be quite a bit larger still.
A strength of this analysis is that anticoagulant reversal trials offer a degree of
rigor for secondary events such as thrombotic complications, as these are carefully
captured primary safety events. Unlike other retrospective analyses using medical
records or databases, the present analysis uses patient-level data gathered prospectively
with independent academic adjudication of thrombotic events and an equal follow-up
period.[9] Retrospective hospital medical record analyses are more difficult as patients are
hospitalized for varying periods creating uneven surveillance (i.e., whether a patient
had an event after discharge may not be known).
Selection bias would have influenced our results—since clinicians made a decision
to restart based on the patient's individual characteristics. Clinicians were less
likely to restart anticoagulation in older patients and those with central nervous
system bleeding. Although we adjusted for these variables in our analysis, we cannot
rule out residual confounding. For example, we have no insight into the factors clinicians
considered in the restart decision, such as thrombotic risk or patient preference,
and we have no data on medication compliance in those who were restarted. One can
infer some restart clinical reasoning from the structure of the models. In the landmark
models, patients with events prior to the landmark time point were excluded, so these
restart decisions were for prevention of thrombosis. The Cox model compares events
per time restarted and not restarted, so it would include restarting both for prevention
and treatment of thrombotic disease (treatment meaning the patient has already had
a thrombotic event post major bleed forcing the restart decision). However, our data
are merely hypothesis-generating and cannot be considered reliable enough to guide
clinical practice.
We excluded bleeding events in the first 3 days. This exclusion is supported by the
natural history of anticoagulant-associated bleeding, particularly in the central
nervous system. Hawryluk et al[1] extracted data from 63 case reports and series on 492 anticoagulated central nervous
system hemorrhage patients, finding a rebleeding rate of 8% and a thrombotic event
rate of 6%. Temporal mapping showed most rebleeding events occurred in the first 72 hours,
while most thrombotic events occurred from 3 days to 1 week after the patient's presentation
with bleeding. Including these early bleeding events, nearly all of which were extensions
of the index bleeding, clouds the model with regard to restarting.
Restart timing is an important and as yet unanswered question. Current retrospective
evidence, expert opinion, and clinical practice on timing are highly variable. Recommendations
range from 3 days[1] to 30 weeks,[2] though the American Stroke Association guideline states ≥1 week with multiple caveats.[21] Majeed et al[2] retrospectively evaluated 234 ICH patients, 59 of whom restarted warfarin. Cox models
found a restart interval of 10 to 30 weeks minimized a composite outcome. Our time-dependent
Cox model attempts to simulate timing's influence by comparing time restarted versus
not in terms of the outcomes and produces an adjusted HR ([Table 3]). The landmark approach is an attempt to incorporate timing by setting points by
which a certain number of (event-free) patients have been restarted and comparing
outcomes thereafter. This can be represented as a Kaplan–Meier curve ([Fig. 2]). It is not the same thing as finding that a time point is a “best” restart time.
This can only be accomplished in randomized models comparing different timing strategies
(such as the Restart TICrH trial[20]) in which each group has a specified restart day. Patients in our dataset could
be restarted at any time in the 30-day follow-up. The more specific one gets to a
certain time interval in the dataset, the less restarted patients there are and thus
the less power to detect differences in outcomes. This probably explains why the 14-day
landmark had the peak power to detect differences. It was enough time for enough patients
to be restarted to have a larger group for comparison with sufficient time remaining
for outcome differences in groups to accumulate and potentially differ. Later landmark
time points reduce power in a slightly different way. The data terminate at 30 days,
so a restart landmark at 3 weeks, for example, allows only 1 week for outcomes to
accumulate and potentially differ. Less outcomes lead to smaller differences between
groups and thus less power to detect those differences.
It is important to categorize these findings in terms of the types of bleeding, i.e.,
spontaneous intracerebral hemorrhage versus traumatic ICH versus GI and other major
hemorrhage. Each has very different pathophysiologies and risks and consequences of
rebleeding with restart of anticoagulation, i.e., rebleeding in the cranium is life-threatening
while rebleeding in the gut may not be. Whether and when to restart must account for
these variables, and this analysis cannot speak to them as we include patients from
all three categories.
Our findings modestly support an emerging body of literature that restarting may confer
a benefit to patients. Park et al[22] retrospectively analyzed 428 atrial fibrillation patients with a history of ICH,
finding benefit in a composite outcome for restarting at ≥2 weeks. It is important
to note that Park et al followed patients for >3 years on average while our study
was limited to 30 days, during which it is more difficult to detect differences in
relatively infrequent events.
This analysis has important limitations. Because all patients received andexanet,
we cannot speak to any thrombotic risk associated with this agent, so we cannot be
sure these findings would translate to patients who did not receive andexanet. Our
analysis is underpowered to detect the likely differences in thrombotic events and
bleeding at restart intervals across the 30-day follow-up period. Since this secondary
analysis is limited to data from ANNEXA-4, there could be uncollected data points
that influence bleeding, thrombosis, death, and/or restart behavior. Posthoc analyses
have known sources of bias (e.g., survivorship and selection, and unknown/uncollected
confounders) that temper our findings. Bleeding events were drawn from investigator-reported
SAE files, which may not include all bleeding complications, though should include
all serious ones. We only found significant associations when we treated all index
major bleedings as a single group, e.g., the ICH subgroup analyses had no significant
findings. Severity of ICH and its surrogates such as hematoma volume likely influence
restart decisions, but we did not detect it in these models. Also, spontaneous ICH,
traumatic ICH, GI, and other bleeding types all have differing risks. This likely
influences the odds and timing of restarting decisions and secondary outcomes.
Conclusion
Restarting therapeutic oral anticoagulation after major hemorrhage was associated
with decreased risk of thrombotic events and increased risk of bleeding in this highly
selected ANNEXA-4 clinical trial population. There is modest evidence for a net benefit
when combining thrombosis, hemorrhage, and death into a composite. These results should
be viewed cautiously as hypothesis-generating, as they are retrospective, and the
reasons for restarting patients could introduce bias. They may support the conduct
of randomized clinical trials to test if restarting anticoagulation after acute major
bleeding has a net clinical benefit.
What is known about this topic?
What does this paper add?
-
In a posthoc analysis of a prospective, 352-patient anticoagulant reversal trial,
no thrombotic events occurred after restarting anticoagulation, but bleeding increased,
and restarting was associated with a decrease in a composite of thrombotic events,
rebleeding, and death, suggesting a net benefit.
