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The New England Journal of Medicine

Original Article
Volume 345:1444-1451

November 15, 2001

Number 20
A Comparison of Warfarin and Aspirin for the Prevention of Recurrent
Ischemic Stroke
J.P. Mohr, M.D., J.L.P. Thompson, Ph.D., R.M. Lazar, Ph.D., B. Levin, M.D.,
R.L. Sacco, M.D., K.L. Furie, M.D., J.P. Kistler, M.D., G.W. Albers, M.D.,
L.C. Pettigrew, M.D., H.P. Adams, Jr., M.D., C.M. Jackson, M.D., P.
Pullicino, M.D., for the Warfarin–Aspirin Recurrent Stroke Study Group
ABSTRACT
Background Despite the use of antiplatelet agents, usually aspirin, in
patients who have had an ischemic stroke, there is still a substantial rate
of recurrence. Therefore, we investigated whether warfarin, which is
effective and superior to aspirin in the prevention of cardiogenic embolism,
would also prove superior in the prevention of recurrent ischemic stroke in
patients with a prior noncardioembolic ischemic stroke.
Methods In a multicenter, double-blind, randomized trial, we compared the
effect of warfarin (at a dose adjusted to produce an international
normalized ratio of 1.4 to 2.8) and that of aspirin (325 mg per day) on the
combined primary end point of recurrent ischemic stroke or death from any
cause within two years.
Results The two randomized study groups were similar with respect to
base-line risk factors. In the intention-to-treat analysis, no significant
differences were found between the treatment groups in any of the outcomes
measured. The primary end point of death or recurrent ischemic stroke was
reached by 196 of 1103 patients assigned to warfarin (17.8 percent) and 176
of 1103 assigned to aspirin (16.0 percent; P=0.25; hazard ratio comparing
warfarin with aspirin, 1.13; 95 percent confidence interval, 0.92 to 1.38).
The rates of major hemorrhage were low (2.22 per 100 patient-years in the
warfarin group and 1.49 per 100 patient-years in the aspirin group). Also,
there were no significant treatment-related differences in the frequency of
or time to the primary end point or major hemorrhage according to the cause
of the initial stroke.
Conclusions Over a two-year period, we found no difference between aspirin
and warfarin in the prevention of recurrent ischemic stroke or death or in
the rate of major hemorrhage. Consequently, we regard both warfarin and
aspirin as reasonable therapeutic alternatives.
  _____

Long-standing doubts, expressed as late as the 1980s, about the efficacy of
warfarin for the prevention of stroke 1
<http://content.nejm.org/cgi/content/full/345/20/#R1>  were mitigated by the
results of more recent clinical trials. Recurrence rates were lower with
warfarin than with placebo in patients who had stroke after myocardial
infarction. 2 <http://content.nejm.org/cgi/content/full/345/20/#R2>  The
rates of first stroke in patients with atrial fibrillation were lower with
warfarin than with a range of other therapies, 3
<http://content.nejm.org/cgi/content/full/345/20/#R3>  placebo, 4
<http://content.nejm.org/cgi/content/full/345/20/#R4>  or aspirin. 5
<http://content.nejm.org/cgi/content/full/345/20/#R5>  Also, in open-label
studies, the rates of recurrent stroke were lower with warfarin than with
placebo or aspirin. 6 <http://content.nejm.org/cgi/content/full/345/20/#R6>
Rates of adverse events with warfarin were acceptably low at the ranges of
the international normalized ratio (INR) used in the studies (1.5 to 3.0). 5
<http://content.nejm.org/cgi/content/full/345/20/#R5> , 7
<http://content.nejm.org/cgi/content/full/345/20/#R7>
Most previous clinical trials of drugs to prevent recurrent ischemic stroke
after a noncardiogenic ischemic stroke studied one or more of a wide variety
of platelet-antiaggregant drugs, particularly aspirin, with which the
recurrence rate approximates 8 percent. 8
<http://content.nejm.org/cgi/content/full/345/20/#R8> , 9
<http://content.nejm.org/cgi/content/full/345/20/#R9> , 10
<http://content.nejm.org/cgi/content/full/345/20/#R10> , 11
<http://content.nejm.org/cgi/content/full/345/20/#R11>  The organizers of
the current trial believed that a trial comparing warfarin and aspirin in
the prevention of recurrent ischemic stroke was justified. This belief was
based on the success of warfarin in the prevention of strokes among patients
with atrial fibrillation and the inference that some ischemic strokes are
due to embolism. 12 <http://content.nejm.org/cgi/content/full/345/20/#R12>
Furthermore, no trial had determined whether anticoagulant agents were
superior to platelet-antiaggregant drugs in preventing other,
noncardioembolic forms of ischemic stroke.
Methods
Study Design
The Warfarin–Aspirin Recurrent Stroke Study (WARSS) was an
investigator-initiated, randomized, double-blind, multicenter clinical trial
conducted in 48 academic medical centers in the United States and sponsored
by the National Institute of Neurological Disorders and Stroke. It also
served as the basis for four parallel stroke studies. 13
<http://content.nejm.org/cgi/content/full/345/20/#R13>  The trial was
formulated and designed by the stroke research staff at the Neurological
Institute of Columbia Presbyterian Medical Center. Clinical data were
collected and monitored by the data-management center in the Stroke Unit at
the Neurological Institute. Management of data on anticoagulant therapy,
double-blinding procedures, and statistical analysis were conducted by the
statistical-analysis center of the Department of Biostatistics, Mailman
School of Public Health, Columbia University. Study medications were
bottled, packaged, and distributed by Quintiles (Mount Laurel, N.J.). To
eliminate variations between laboratories, 14
<http://content.nejm.org/cgi/content/full/345/20/#R14>  blood samples for
determination of the INR were processed centrally by Quest Diagnostics
(Teterboro, N.J.). The study protocol was approved by the institutional
review board at each participating center. Written informed consent was
obtained from each patient. Patient recruitment began in June 1993, and
follow-up ended, as scheduled, in June 2000.
Eligibility
Eligible patients were 30 to 85 years old, were considered acceptable
candidates for warfarin therapy, had had an ischemic stroke within the
previous 30 days, and had scores of 3 or more on the Glasgow Outcome Scale.
On this scale a score of 3 indicates severe disability, a score of 4
moderate disability, and a score of 5 minimal or no disability. Patients
were ineligible if they had a base-line INR above the normal range (more
than 1.4), stroke that was due to a procedure or that was attributed to
high-grade carotid stenosis for which surgery was planned, or stroke
associated with an inferred cardioembolic source; most of the last group had
atrial fibrillation at the time of stroke. Eligibility was verified before
randomization by telephone contact with the data-management center, in which
each criterion for eligibility or ineligibility, the dates of stroke and
randomization, magnetic resonance imaging or computed tomography of the
brain, and signing of the consent form were confirmed.
Medications and Blinding
The medications evaluated were aspirin (Bayer, Morristown, N.J.), one 325-mg
tablet daily, and warfarin (Dupont, Wilmington, Del.), one 2-mg scored
tablet daily. The warfarin doses were adjusted to achieve and maintain an
INR in the range of 1.4 to 2.8. The patients were randomly assigned to
receive active aspirin and warfarin placebo or active warfarin and aspirin
placebo. Randomization was stratified according to site. No patients
received two placebos or two active treatments. All centers and patients
were informed as to the double-blind design and the plan for the use of
false INR values in the group receiving active aspirin and warfarin placebo.
All centers followed the same schedule of visits to the clinic for drawing
of blood to measure the INR, monitoring of medication, and adjustment of the
dose of warfarin or warfarin placebo.
Blood samples for determination of the INR were sent to Quest Diagnostics on
the same day or by overnight courier service. Before a center was admitted
as a study site, we confirmed that blood samples sent to Quest Diagnostics
were viable and yielded reliable INR determinations. All INR results were
transferred electronically to the statistical-analysis center, which sent
the results to the local centers by facsimile transmission. According to
prior agreement among the center clinicians and with the use of a method
validated early in the trial, 15
<http://content.nejm.org/cgi/content/full/345/20/#R15>  the INR results sent
to local centers were unmodified for the patients receiving active warfarin,
but for patients receiving active aspirin and warfarin placebo, they were
replaced by the statistical-analysis center with fabricated values that were
plausible for the dose and duration of warfarin therapy. No INR results were
available directly to the local centers from Quest Diagnostics. According to
the guidelines of the Food and Drug Administration, high INR values (4.5 or
more) were forwarded to the data-management center and transmitted
immediately to local centers by cellular telephone. To preserve blinding,
some emergency notifications for falsely elevated values in patients
receiving warfarin placebo were also sent by the statistical-analysis
center. The principal clinical investigator reviewed all outgoing INR
reports, writing a personal cautionary note to the local investigator in the
case of reports showing trends for values below or above the desired ranges.
All participants other than the principal statistical investigator at the
statistical-analysis center were blinded to the patients' study-group
assignments. During the course of the trial, unblinding was required for 15
patients, in most cases because of the need for an invasive surgical
procedure. All 15 patients stopped treatment with study drugs, but their
data were included in the intention-to-treat analysis.
Follow-up
Patients were followed for 2 years ±1 month, up to a maximum of 761 days.
Follow-up was conducted monthly by telephone or in person at the time of
drawing of blood for the determination of the INR to assess compliance and
to regulate INR values, quarterly in person for clinical evaluation, and
annually for detailed examination; the occurrence of end points was also
ascertained at each contact. Personnel at the data-management center also
conducted site visits to audit the records of all patients at each center
for end points and adverse events.
Assessment of End Points and Major Adverse Events
The primary end point was death from any cause or recurrent ischemic stroke,
whichever occurred first. Recurrent ischemic stroke was defined as a new
lesion detected by computed tomography or magnetic resonance imaging or, in
the absence of a new lesion, clinical findings consistent with the
occurrence of stroke that lasted for more than 24 hours. Local centers
reported potential outcome events to the events coordinator at the
data-management center and submitted clinical summaries, study forms
documenting clinical details, and brain imaging studies. An independent,
treatment-blinded neuroradiologist reviewed the images. Five
treatment-blinded neurologists adjudicated all clinical events using a
majority verdict for decisions about outcomes.
Major hemorrhage was defined as intracranial, intraspinal, intracerebral,
subarachnoid, subdural, or epidural hemorrhage or any other bleeding event
requiring transfusion. Minor hemorrhage, which did not require transfusion,
included gastrointestinal, genitourinary, retroperitoneal, joint,
subcutaneous or muscular, gingival or oral, and conjunctival hemorrhage;
epistaxis; hemoptysis; ecchymoses; and hemorrhage after trauma or from
multiple sites. A treatment-blinded adjudicator classified hemorrhagic
events as major or minor, reviewed data on death due to any reported
hemorrhage, and determined the relation of the hemorrhage to treatment.
Statistical Analysis
The primary null hypothesis was that there would be no difference between
patients receiving warfarin and those receiving aspirin in the time to or
rate of death from any cause or recurrent ischemic stroke. Secondary null
hypotheses of major clinical interest were that there would be no
differences in the time to either component of the primary end point or to
major hemorrhage according to sex, race or ethnic group, or cause of prior
stroke.
The original target sample size was 1920 patients, which provided the study
with 80 percent power and a 5 percent two-sided probability of a type I
error for a test of the primary null hypothesis according to the intention
to treat, allowing for a 30 percent reduction in the event rate for one
therapy from a 16 percent event rate over two years for the other, and an
overall dropout and discontinuation rate of 20 percent at two years for both
therapies combined. In 1995, while still blinded to event rates according to
treatment group, the performance and safety monitoring board appointed by
the National Institute of Neurological Disorders and Stroke increased the
target sample size to 2200 to adjust for the possible effects of
interruption of therapy. In 1996, they revised the original stopping rule
based on a single interim analysis by adopting a modified repeated
significance test 16 <http://content.nejm.org/cgi/content/full/345/20/#R16>
procedure that called for three scheduled interim analyses and allowed for
additional interim analyses. The trial proceeded to its planned completion
and final analysis without crossing the efficacy or safety boundaries.
All the major study hypotheses were prespecified and tested on an
intention-to-treat basis with a two-tailed alpha of 0.05. The Kaplan–Meier
method 17 <http://content.nejm.org/cgi/content/full/345/20/#R17>  was used
to estimate curves for the length of time to the event, and the log-rank
test 18 <http://content.nejm.org/cgi/content/full/345/20/#R18>  was used to
compare the cumulative incidence curves in the treatment groups. The primary
analysis was adjusted for loss to follow-up by a prespecified stratified
imputation procedure that distinguishes different types of loss to follow-up
and incorporates assumptions appropriate to each. The reported P values and
confidence intervals have not been adjusted for interim analyses.
Results
A total of 2206 patients were randomly assigned to treatment groups at a
steady rate during the recruitment phase. Their clinical and demographic
features are shown in Table 1
<http://content.nejm.org/cgi/content/full/345/20/#T1> . Of these, 1302 (59
percent) were over the age of 60 years, 1309 (59 percent) were male, 1499
(68 percent) had hypertension, 705 (32 percent) had diabetes, 504 (23
percent) had cardiac disease, 390 (18 percent) had angina or prior
myocardial infarction, and 629 (29 percent) had prior amaurosis fugax,
transient ischemic attack, or stroke. The end-point status at two years was
established for 2173 (98.5 percent). An additional 33 (1.5 percent) withdrew
consent or were lost to follow-up for other reasons, at a mean of 10.2±7.5
months after randomization. Figure 1
<http://content.nejm.org/cgi/content/full/345/20/#F1>  illustrates follow-up
and imputation of events according to treatment.


View this table:
[in this window] <http://content.nejm.org/cgi/content/full/345/20/1444/T1>
[in a new window]
<http://content.nejm.org/cgi/content-nw/full/345/20/1444/T1>

Table 1. Base-Line Characteristics of the Patients.



  <http://content.nejm.org/cgi/content/full/345/20/1444/F1>
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Figure 1. Follow-up of Patients and Imputation of Events.
Events for which exact dates were unknown were considered to have occurred
at the midpoints of the calendar periods during which they occurred.
Plus–minus values are means ±SE.

Laboratory Testing
Quest Diagnostics determined 48,931 INR values. The mean interval between
the dates of blood sampling was 27.9±12.6 days. The mean daily INR for
patients taking warfarin was 2.1 (median, 1.9). Overall, 70.7 percent of
daily INR values determined 28 or more days after randomization were within
the target range (1.4 to 2.8), 13.0 percent were above the range, and 16.3
percent were below the range. There were no significant differences in INR
values among patients with different types of prior stroke (cryptogenic;
small-vessel or lacunar; severe stenosis, or occlusion of a large artery;
other, determined cause; and conflicting mechanism [there was more than one
diagnostic possibility]) (P=0.24 by F test with log-transformed INR values).
Outcomes
The overall rate of the primary end point of death or recurrent ischemic
stroke of 16.9 percent (372 of 2206 patients) slightly exceeded the 16
percent rate assumed in the trial design. In the primary intention-to-treat
analysis, there were no significant differences between the warfarin and
aspirin groups in the time to the primary end point (P=0.25 by two-tailed
log-rank test; hazard ratio for warfarin as compared with aspirin, 1.13; 95
percent confidence interval, 0.92 to 1.38; two-year probability of an event,
17.8 percent with warfarin and 16.0 percent with aspirin) ( Table 2
<http://content.nejm.org/cgi/content/full/345/20/#T2>  and Figure 2
<http://content.nejm.org/cgi/content/full/345/20/#F2> ). Censoring data from
subjects whose data were incomplete at the time of loss to follow-up did not
materially affect the outcome of the primary analysis, and incorporating the
interruption of study medication as a time-dependent covariate showed that
the effects of warfarin and aspirin therapy did not differ.


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[in this window] <http://content.nejm.org/cgi/content/full/345/20/1444/T2>
[in a new window]
<http://content.nejm.org/cgi/content-nw/full/345/20/1444/T2>

Table 2. Results of Primary and Secondary Analyses.



  <http://content.nejm.org/cgi/content/full/345/20/1444/F2>
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Figure 2. Kaplan–Meier Analyses of the Time to Recurrent Ischemic Stroke or
Death According to Treatment Assignment.

The rates of major hemorrhage were low, with no significant differences
between treatment groups; the annual rates were 2.22 per 100 patient-years
for warfarin and 1.49 per 100 patient-years for aspirin (rate ratio, 1.48;
P=0.10). Patients in the warfarin group had significantly more minor
hemorrhages than did those in the aspirin group ( Table 3
<http://content.nejm.org/cgi/content/full/345/20/#T3> ). There was no
significant difference between groups in the time to the first occurrence of
major hemorrhage or the primary end point (P=0.16; hazard ratio with
warfarin as compared with aspirin, 1.15; 95 percent confidence interval,
0.95 to 1.39) ( Table 2
<http://content.nejm.org/cgi/content/full/345/20/#T2> ).


View this table:
[in this window] <http://content.nejm.org/cgi/content/full/345/20/1444/T3>
[in a new window]
<http://content.nejm.org/cgi/content-nw/full/345/20/1444/T3>

Table 3. Adverse Events According to Treatment Assignment.

There were also no significant differences in the time to a primary end
point between patients of different sexes, of different racial or ethnic
groups, or with different types of prior stroke ( Table 2
<http://content.nejm.org/cgi/content/full/345/20/#T2> ). Figure 3
<http://content.nejm.org/cgi/content/full/345/20/#F3>  shows INR-specific
rates of primary events plotted by the method of Rosendaal et al., 19
<http://content.nejm.org/cgi/content/full/345/20/#R19>  with use of the last
INR value before the event. The rates decline for INR values until the INR
interval of 1.5 to less than 2.0, but change little thereafter.


  <http://content.nejm.org/cgi/content/full/345/20/1444/F3>
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Figure 3. Incidence of Recurrent Ischemic Stroke or Death among Patients
Assigned to Warfarin, According to the International Normalized Ratio.
The I bars indicate 95 percent confidence intervals. The international
normalized ratio was the last measured before the event.

Discussion
We observed no significant difference between treatment with warfarin and
treatment with aspirin in the prevention of recurrent ischemic stroke or
death or in the occurrence of serious adverse events in this large cohort of
patients with inferred noncardioembolic ischemic stroke. Not only did the
use of warfarin not lead to a 30 percent reduction in the risk of recurrent
stroke — the reduction used to estimate the sample size — but it was also
associated with a nonsignificant, 13 percent higher increase in risk over
that with aspirin. Treatment with warfarin did not result in excess event
rates during the first 30 days or in a significant increase in the rates of
hemorrhage; these potential outcomes affected the trial design because of
concern that either of these outcomes would offset any benefit of warfarin.
Two observations suggest that the demographic characteristics of the study
population and outcomes compare favorably with those of other trials of
aspirin or warfarin. First, the event rate among patients assigned to
aspirin was similar to that in other trials of aspirin for the prevention of
recurrent ischemic stroke. 10
<http://content.nejm.org/cgi/content/full/345/20/#R10> , 11
<http://content.nejm.org/cgi/content/full/345/20/#R11> , 12
<http://content.nejm.org/cgi/content/full/345/20/#R12> , 14
<http://content.nejm.org/cgi/content/full/345/20/#R14>  Furthermore, the low
rates of hemorrhage with warfarin were similar to those in warfarin-treated
patients with stroke associated with atrial fibrillation whose INR values
were similar to those of our patients. 5
<http://content.nejm.org/cgi/content/full/345/20/#R5> , 7
<http://content.nejm.org/cgi/content/full/345/20/#R7>  Our finding that the
rate of recurrent stroke with warfarin was similar to the rate with aspirin
suggests that warfarin is an effective therapy in patients with a prior
ischemic stroke. However, in our trial, warfarin was not superior to
aspirin. If anything, the reverse was true; warfarin did not decrease the
rate of severe recurrent stroke, as it does in patients with prior stroke
associated with atrial fibrillation. 5
<http://content.nejm.org/cgi/content/full/345/20/#R5> , 6
<http://content.nejm.org/cgi/content/full/345/20/#R6> , 7
<http://content.nejm.org/cgi/content/full/345/20/#R7>  Moreover, warfarin
costs more than aspirin and requires close monitoring.
It is unlikely that the range of INR values chosen was too low to show the
superiority of warfarin. Treatment targeted to the same range of values was
successful for the prevention of first strokes in patients with atrial
fibrillation. Published graphs showing the effect of the INR on the risk of
stroke showed curves similar in shape to those in our results, flattening
for INR values of 1.5 to 2.0 and remaining relatively stable for higher
values up to 3.0. However, the event rates in relation to the same range of
INR values (1.5 to 3.0) among patients with atrial fibrillation were well
below that in our study. 5
<http://content.nejm.org/cgi/content/full/345/20/#R5> , 6
<http://content.nejm.org/cgi/content/full/345/20/#R6> , 7
<http://content.nejm.org/cgi/content/full/345/20/#R7>
We considered using higher INR values than those used in studies of patients
with atrial fibrillation, but observations from other studies published
during the course of our study supported our concern about safety. 20
<http://content.nejm.org/cgi/content/full/345/20/#R20> , 21
<http://content.nejm.org/cgi/content/full/345/20/#R21> , 22
<http://content.nejm.org/cgi/content/full/345/20/#R22> , 23
<http://content.nejm.org/cgi/content/full/345/20/#R23> , 24
<http://content.nejm.org/cgi/content/full/345/20/#R24>  Higher rates of
major hemorrhage could have stopped the trial before efficacy could be
validly tested, as happened for the Stroke Prevention in Reversible Ischemia
Trial, an open-label comparison of warfarin with lower-dose aspirin after
transient ischemic attacks and stroke that used an INR range of 3.0 to 4.5
(mean, 3.5). 25 <http://content.nejm.org/cgi/content/full/345/20/#R25>
Higher INR ranges than those we used in other, nonstroke settings have had
mixed results with respect to safety as compared with studies of warfarin
alone 26 <http://content.nejm.org/cgi/content/full/345/20/#R26>  or in
combination with aspirin. 27
<http://content.nejm.org/cgi/content/full/345/20/#R27>
The overall percentages of patients with INR values in, above, or below the
target range in our study also compare favorably with the percentages in
other trials. These findings argue against the possibility that warfarin's
lack of superiority to aspirin was due to high percentages of patients with
low INR values. Because reports of studies showing the success of warfarin
in patients with atrial fibrillation did not present data on the time course
of INR values during the trials in graphic form, no direct time-based
comparisons with our data are possible.
As a direct test of warfarin versus aspirin for the prevention of recurrent
ischemic stroke in a broad clinical setting (excluding patients with stroke
due to embolism), our study necessarily included patients with a variety of
types of prior ischemic stroke. Because it is not always easy to separate
different types of stroke, regardless of the classification scheme used, 28
<http://content.nejm.org/cgi/content/full/345/20/#R28> , 29
<http://content.nejm.org/cgi/content/full/345/20/#R29> , 30
<http://content.nejm.org/cgi/content/full/345/20/#R30>  some patients with
cardiogenic embolism may have been included. If so, they did not favorably
affect the findings with regard to the effect of warfarin. The recurrence
rates in patients with different types of prior ischemic stroke are similar
to those found in the Stroke Data Bank of the National Institute of
Neurological Disorders and Stroke 28
<http://content.nejm.org/cgi/content/full/345/20/#R28>  and the Northern
Manhattan Stroke Study 31
<http://content.nejm.org/cgi/content/full/345/20/#R31>  but differ somewhat
from those in other studies. 32
<http://content.nejm.org/cgi/content/full/345/20/#R32>
Like the studies of tissue plasminogen activator for acute stroke, 33
<http://content.nejm.org/cgi/content/full/345/20/#R33>  our study did not
find significant differences in the effects of treatment among patients with
different clinically identifiable types of prior ischemic stroke. Despite
our study's lack of sufficient power to show such differences, our data
nonetheless suggest some possible selective treatment effects. Aspirin was
slightly, but not significantly, superior to warfarin in patients with
large-vessel and lacunar infarcts. Patients with large-vessel strokes are
currently under study. 34
<http://content.nejm.org/cgi/content/full/345/20/#R34>  If aspirin is
superior to warfarin in lacunar stroke, that finding will support the idea
that there is a mechanistic link between lacunar disease and
large-intracranial-artery atheroma. 35
<http://content.nejm.org/cgi/content/full/345/20/#R35> , 36
<http://content.nejm.org/cgi/content/full/345/20/#R36>  Cryptogenic stroke,
in which the prevalence of superficial brain convexity infarcts and lack of
evidence of large-artery disease have made clinically occult embolism 15
<http://content.nejm.org/cgi/content/full/345/20/#R15>  or coagulopathy 37
<http://content.nejm.org/cgi/content/full/345/20/#R37>  the leading presumed
causes, was the only clinically identified stroke type for which a possible
benefit of warfarin was suggested by our data; but the reduction in risk was
small (8 percent) and not statistically significant.
Warfarin offered no additional benefit over aspirin in preventing recurrent
ischemic stroke in the population we studied. Patients with other,
established reasons for warfarin use may take comfort in the evidence of
safety and lack of significant difference overall, as compared with aspirin.
However, aspirin, either alone or in combination with some other
antiplatelet agents, 38
<http://content.nejm.org/cgi/content/full/345/20/#R38>  appears to be a
well-justified choice for the prevention of recurrent ischemic stroke.
<http://weeklybriefings.org/feature.asp?strXmlDoc=3452002>
Supported by a grant (RO1-NS-28371) from the National Institute of
Neurological Disorders and Stroke. Medications and placebos were supplied by
Dupont Pharmaceuticals and Bayer.
* Participants in the study group are listed in the Appendix.
<http://content.nejm.org/cgi/content/full/345/20/#RFN1>

Source Information
From the Neurological Institute (J.P.M., R.M.L., R.L.S.) and the Department
of Biostatistics (J.L.P.T., B.L.), Columbia Presbyterian Medical Center, New
York; Massachusetts General Hospital, Boston (K.L.F., J.P.K.); Stanford
University Medical Center, Palo Alto, Calif. (G.W.A.); the University of
Kentucky Medical Center, Louisville (L.C.P.); University of Iowa Health
Care, Iowa City (H.P.A.); the University of California at San Diego, San
Diego (C.M.J.); and the State University of New York at Buffalo, Buffalo
(P.P.).
Address reprint requests to Dr. Mohr at the Neurological Institute, 710 W.
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References
1.      Sandercock P, Warlow C, Bamford J, Peto R, Starkey I. Is a controlled
trial of long-term oral anticoagulants in patients with stroke and
non-rheumatic atrial fibrillation worthwhile? Lancet 1986;1:788-792.
[Medline]
<http://content.nejm.org/cgi/external_ref?access_num=2870278&link_type=MED>
2.      A double-blind trial to assess long-term oral anticoagulant therapy in
elderly patients after myocardial infarction: report of the Sixty Plus
Reinfarction Study Research Group. Lancet 1980;2:989-994. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=6107674&link_type=MED>
3.      The Boston Area Anticoagulation Trial for Atrial Fibrillation
Investigators. The effect of low-dose warfarin on the risk of stroke in
patients with nonrheumatic atrial fibrillation. N Engl J Med
1990;323:1505-1511. [Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=nejm&resid=323
/22/1505>
4.      Ezekowitz MD, Bridgers SL, James KE, et al. Warfarin in the prevention of
stroke associated with nonrheumatic atrial fibrillation. N Engl J Med
1992;327:1406-1412. [Erratum, N Engl J Med 1993;328:148.] [Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=nejm&resid=327
/20/1406>
5.      Stroke Prevention in Atrial Fibrillation Investigators. Warfarin versus
aspirin for prevention of thromboembolism in atrial fibrillation: Stroke
Prevention in Atrial Fibrillation II Study. Lancet 1994;343:687-691.
[Medline]
<http://content.nejm.org/cgi/external_ref?access_num=7907677&link_type=MED>
6.      Petersen P, Boysen G, Godtfredsen J, Andersen ED, Andersen B.
Placebo-controlled, randomised trial of warfarin and aspirin for prevention
of thromboembolic complications in chronic atrial fibrillation: the
Copenhagen AFASAK study. Lancet 1989;1:175-179. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=2563096&link_type=MED>
7.      Goldstein LB, Adams R, Becker K, et al. Primary prevention of ischemic
stroke: a statement for healthcare professionals from the Stroke Council of
the American Heart Association. Stroke 2001;32:280-299. [Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=FULL&journalCode=strokeaha&resi
d=32/1/280>
8.      Bousser MG, Eschwege E, Hageunau M, et al. "AICLA" controlled trial of
aspirin and dipyridamole in the secondary prevention of athero-thrombotic
cerebral ischemia. Stroke 1983;14:5-14. [Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=strokeaha&resi
d=14/1/5>
9.      Hass WK, Easton JD, Adams HP Jr, et al. A randomized trial comparing
ticlopidine hydrochloride with aspirin for the prevention of stroke in
high-risk patients. N Engl J Med 1989;321:501-507. [Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=nejm&resid=321
/8/501>
10.     CAPRIE Steering Committee. A randomised, blinded, trial of clopidogrel
versus aspirin in patients at risk of ischaemic events (CAPRIE). Lancet
1996;348:1329-1339. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=8918275&link_type=MED>
11.     European Stroke Prevention Study 2: efficacy and safety data. J Neurol
Sci 1997;151:Suppl:S1-S77. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=9276859&link_type=MED>
12.     Mohr JP. Cryptogenic stroke. N Engl J Med 1988;318:1197-1198. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=3362167&link_type=MED>
13.     The WARSS, APASS, PICSS, HAS, and GENESIS Study Groups. The feasibility
of a collaborative double-blind study using an anticoagulant. Cerebrovasc
Dis 1997;7:100-112.
14.     Lind SE, Pearce LA, Feinberg WM, Bovill EG. Clinically significant
differences in the International Normalized Ratio measured with reagents of
different sensitivities. Blood Coagul Fibrinolysis 1999;10:215-227.
[Medline]
<http://content.nejm.org/cgi/external_ref?access_num=10456611&link_type=MED>
15.     Thompson JLP, Fleiss JL, James K, et al. Test of an algorithm for
simulating prothrombin times in a double-blind anticoagulation drug trial.
Ann Neurol 1994;36:305-306.abstract
16.     Modified repeated significance tests. In: Siegmund D. Sequential
analysis: tests and confidence intervals. New York: Springer-Verlag,
1985:86-9.
17.     Kaplan EL, Meier P. Nonparametric estimation from incomplete
observations. J Am Stat Assoc 1958;53:457-81.
18.     Mantel N. Evaluation of survival data and two new rank order statistics
arising in its consideration. Cancer Chemother Rep 1966;50:163-170.
[Medline]
<http://content.nejm.org/cgi/external_ref?access_num=5910392&link_type=MED>
19.     Rosendaal FR, Cannegieter SC, van der Meer FJM, Briët E. A method to
determine the optimal intensity of oral anticoagulant therapy. Thromb
Haemost 1993;69:236-239. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=8470047&link_type=MED>
20.     Gulløv AL, Koefoed BG, Petersen P. Bleeding during warfarin and aspirin
therapy in patients with atrial fibrillation: the AFASAK 2 study. Arch
Intern Med 1999;159:1322-1328. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=10386508&link_type=MED>
21.     Yamaguchi T. Optimal intensity of warfarin therapy for secondary
prevention of stroke in patients with nonvalvular atrial fibrillation: a
multicenter, prospective, randomized trial. Stroke 2000;31:817-821.
[Abstract/Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=strokeaha&resi
d=31/4/817>
22.     Pengo V, Legnani C, Noventa F, Palareti G. Oral anticoagulant therapy in
patients with nonrheumatic atrial fibrillation and risk of bleeding: a
Multicenter Inception Cohort study. Thromb Haemost 2001;85:418-422.
[Medline]
<http://content.nejm.org/cgi/external_ref?access_num=11307807&link_type=MED>
23.     Liu M, Counsell C, Sandercock P. Anticoagulants for preventing
recurrence following ischaemic stroke or transient ischemic attack (Cochrane
review). Cochrane Library, issue 3. Oxford, England: Update Software, 2001.
(Accessed September 18, 2001, at
http://www.update-software.com/abstracts/ab000248.htm.)
24.     Go AS, Hylek EM, Phillips KA, et al. Implications of stroke risk
criteria on the anticoagulation decision in nonvalvular atrial fibrillation:
the Anticoagulation and Risk Factors in Atrial Fibrillation (ATRIA) study.
Circulation 2000;102:11-13. [Abstract/Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=circulationaha
&resid=102/1/11>
25.     Gorter JW. Major bleeding during anticoagulation after cerebral
ischemia: patterns and risk factors: Stroke Prevention in Reversible
Ischemia Trial (SPIRIT). Neurology 1999;53:1319-1327. [Abstract/Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=neurology&resi
d=53/6/1319>
26.     Witte K, Thackray S, Clark AL, Cooklin M, Cleland JG. Clinical trials
update: IMPROVEMENT-HF, COPERNICUS, MUSTIC, ASPECT-II, APRICOT and HEART.
Eur J Heart Fail 2000;2:455-460. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=11113724&link_type=MED>
27.     Huynh T, Théroux P, Bogaty P, Nasmith J, Solymoss S. Aspirin, warfarin,
or the combination for secondary prevention of coronary events in patients
with acute coronary syndromes and prior coronary artery bypass surgery.
Circulation 2001;103:3069-3074. [Abstract/Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=circulationaha
&resid=103/25/3069>
28.     Foulkes MA, Wolf PA, Price TR, Mohr JP, Hier DB. The Stroke Data Bank:
design, methods, and baseline characteristics. Stroke 1988;19:547-554.
[Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=strokeaha&resi
d=19/5/547>
29.     Anderson CS, Taylor BV, Hankey GJ, Stewart-Wynne EG, Jamrozik KD.
Validation of a clinical classification for subtypes of acute cerebral
infarction. J Neurol Neurosurg Psychiatry 1994;57:1173-1179. [Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=jnnp&resid=57/
10/1173>
30.     Madden KP, Karanjia PN, Adams HP Jr, Clarke WR. Accuracy of initial
stroke subtype diagnosis in the TOAST study: Trial of ORG 10172 in Acute
Stroke Treatment. Neurology 1995;45:1975-1979. [Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=neurology&resi
d=45/11/1975>
31.     Sacco RL, Shi T, Zamanillo MC, Kargman DE. Predictors of mortality and
recurrence after hospitalized cerebral infarction in an urban community: the
Northern Manhattan Stroke Study. Neurology 1994;44:626-634. [Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=neurology&resi
d=44/4/626>
32.     Petty GW, Brown RD Jr, Whisnant JP, Sicks JD, O'Fallon WM, Wiebers DO.
Ischemic stroke subtypes: a population-based study of functional outcome,
survival, and recurrence. Stroke 2000;31:1062-1068. [Abstract/Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=strokeaha&resi
d=31/5/1062>
33.     Kwiatkowski TG, Libman RB, Frankel M, et al. Effects of tissue
plasminogen activator for acute ischemic stroke at one year. N Engl J Med
1999;340:1781-1787. [Abstract/Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=nejm&resid=340
/23/1781>
34.     The Warfarin-Aspirin Symptomatic Intracranial Disease (WASID) Study
Group. Prognosis of patients with symptomatic vertebral or basilar artery
stenosis. Stroke 1998;29:1389-1392. [Abstract/Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=strokeaha&resi
d=29/7/1389>
35.     Fisher CM. The arterial lesions underlying lacunes. Acta Neuropathol
(Berl) 1968;12:1-15. [Medline]
<http://content.nejm.org/cgi/external_ref?access_num=5708546&link_type=MED>
36.     Kappelle LJ, van Latum JC, van Swieten JC, Algra A, Koudstaal PJ, van
Gijn J. Recurrent stroke after transient ischaemic attack or minor ischaemic
stroke: does the distinction between small and large vessel disease remain
true to type? J Neurol Neurosurg Psychiatry 1995;59:127-131. [Abstract]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=jnnp&resid=59/
2/127>
37.     Bushnell CD, Goldstein LB. Diagnostic testing for coagulopathies in
patients with ischemic stroke. Stroke 2000;31:3067-3078. [Abstract/Full
Text]
<http://content.nejm.org/cgi/ijlink?linkType=ABST&journalCode=strokeaha&resi
d=31/12/3067>
38.     Albers GW, Amarenco P, Easton JD, Sacco RL, Teal P. Antithrombotic and
thrombolytic therapy for ischemic stroke. Chest 2001;119:Suppl:300S-320S.
[Full Text]
<http://content.nejm.org/cgi/ijlink?linkType=FULL&journalCode=chest&resid=11
9/1_suppl/300S>

Edward E. Rylander, M.D.
Diplomat American Board of Family Practice.
Diplomat American Board of Palliative Medicine.