Extracranial-Intracranial Bypass Surgery for Stroke Prevention in Hemodynamic Cerebral Ischemia: The Carotid Occlusion Surgery Study: A Randomized Trial

Abstract

Context

Patients with symptomatic atherosclerotic internal carotid artery occlusion (AICAO) and hemodynamic cerebral ischemia are at high risk for subsequent stroke when treated medically.

Objective

Test the hypothesis that extracranial-intracranial (EC-IC) bypass surgery, added to best medical therapy, reduces subsequent ipsilateral ischemic stroke in patients with recently symptomatic AICAO and hemodynamic cerebral ischemia.

Design

Parallel group, randomized, open-label, blinded-adjudication clinical treatment trial conducted from 2002–2010.

Setting

49 clinical centers and 18 positron emission tomography (PET) centers in the United States and Canada. The majority were academic medical centers.

Participants

Arteriographically-confirmed AICAO causing hemispheric symptoms within 120 days and hemodynamic cerebral ischemia identified by ipsilateral increased oxygen extraction fraction measured by PET. 195 were randomized: 97 to surgery and 98 to no surgery. Follow-up for the primary endpoint until occurrence, 2 years, or end of trial was 99% complete. No participant withdrew because of adverse events.

Interventions

Anastomosis of superficial temporal artery branch to a middle cerebral artery cortical branch for the surgical group. Anti-thrombotic therapy and risk factor intervention were recommended for all.

Main Outcome Measure

For all participants who were assigned to surgery and received surgery, the combination of (1) all stroke and death from surgery through 30 days post surgery and (2) ipsilateral ischemic stroke within 2 years of randomization. For the nonsurgical group and participants assigned to surgery who did not receive surgery was the combination of (1) all stroke and death from randomization to randomization plus 30 days and (2) ipsilateral ischemic stroke within two years of randomization.

Results

The trial was terminated early for futility. Two-year rates for the primary endpoint were 21.0% (95% CI, 12.8% to 29.2%; 20 events) for the surgical group and 22.7% (95% CI, 13.9% to 31.6%; 20 events) for the nonsurgical group (p=0.78, z-test); difference = 1.7% (95% CI, −10.4% to 13.8%). Thirty-day rates for ipsilateral ischemic stroke were 14.3% (14/97) in the surgical group and 2.0% (2/98) in the nonsurgical group; difference = (95% CI, 4.9% to 19.9%)

Conclusions

Among participants with recently symptomatic AICAO and hemodynamic cerebral ischemia, EC-IC bypass surgery plus medical therapy compared to medical therapy alone did not reduce the risk of recurrent ipsilateral ischemic stroke at 2 years.

INTRODUCTION

Atherosclerotic internal carotid artery occlusion (AICAO) causes approximately 10% of transient ischemic attacks (TIAs) and 15–25% of ischemic strokes in the carotid territory.^1,2 The 2-year risk of subsequent ipsilateral ischemic stroke on medical therapy is 10–15%.^1–3 Extracranial-intracranial (EC-IC) arterial bypass surgery was developed to prevent subsequent stroke by improving hemodynamics distal to the occluded artery.⁴ In 1985, a randomized trial demonstrated no benefit of this surgery in 808 patients with symptomatic carotid artery occlusion.⁵ This trial was criticized for failing to identify the subgroup of patients with hemodynamic cerebral ischemia due to poor collateral circulation for whom surgical revascularization might be of greatest benefit.^6–8 Subsequent advances in neuroimaging have made it possible to identify those with hemodynamic cerebral ischemia who are at high risk for subsequent stroke when treated medically.^3,9–11 We conducted the Carotid Occlusion Surgery Study (COSS) to determine if EC-IC bypass surgery, added to best medical therapy, reduces subsequent ipsilateral ischemic stroke at 2 years in patients with recently symptomatic AICAO and hemodynamic cerebral ischemia identified by positron emission tomography (PET) measurements of oxygen extraction fraction (OEF).

METHODS

COSS was a parallel group, 1:1 randomized, open-label, blinded-adjudication treatment trial conducted at 49 clinical centers and 18 PET centers in the United States and Canada. Personnel at the clinical coordinating center, including the principal investigator and project manager, were blinded to treatment assignment and to outcome, except that persons who paid local sites or processed post-operative PET scans knew treatment assignment. Protocol amendments were made during the trial with approval of the data, safety and monitoring board (DSMB) appointed by the National Institute of Neurological Diseases and Stroke (NINDS) (eMethods 1). This description is based on the final protocol (copy available on request).

Potential participants were identified by monitoring vascular imaging studies and soliciting referrals from physicians. Primary clinical inclusion criteria were (1) vascular imaging demonstrating complete occlusion of an internal carotid artery and (2) TIA or ischemic stroke in the hemispheric territory of an occluded internal carotid artery within 120 days. (See eMethods 2 for full listing of clinical eligibility criteria.) Participants who fulfilled initial clinical eligibility criteria provided written informed consent according to local IRB regulations and proceeded to PET. Participants underwent PET at COSS-approved PET sites (United States Food and Drug Administration Investigational New Drug Application #62,657). Forty second emission images were compiled from 2-sec dynamic images obtained following intravenous administration of 75 mCi H₂¹⁵O and following inhalation of 100 mCi O¹⁵O. Ipsilateral-to-contralateral ratios of mean regional carotid territory OEF were calculated at Washington University from a quotient image of O¹⁵O / H₂¹⁵O PET counts.¹¹ A ratio >1.130 was required. Intra-arterial catheter arteriography documenting (1) occlusion of the symptomatic internal carotid artery and (2) intracranial and extracranial arteries suitable for anastomosis was also required.

Baseline clinical evaluation was performed prior to PET. Data on race/ethnicity in fixed categories were collected according to National Institutes of Health clinical trial requirements. Self-reporting or self-identification was the preferred method for collecting data on ethnicity and race. Respondents were offered the option of selecting one or more racial designations.

1:1 randomization sequences using permuted blocks with stratification for clinical site and to insure comparable numbers with contralateral carotid stenosis in each group were generated using the SAS uniform random number generator (RANUNI). Sequences for each clinical site were loaded into a secure part of the COSS SQL server database. When a local investigator received notification of PET eligibility and entered the information from the Initial Eligibility and Arteriography forms, treatment assignment was obtainable from the COSS website.

Surgical intervention

Surgical intervention was microsurgical end-to-side anastomosis of a superficial temporal artery branch to a cortical branch of the middle cerebral artery. If the superficial temporal branch was unsuitable, the occipital artery could be used. Neurosurgeons were certified by (1) attendance at a 2-day training workshop with videotaped instruction and surgical practice of microvascular anastomosis, (2) demonstration of ≥ 80% graft patency and ≤10% stroke and death rate at 1 month in at least 10 consecutive previous EC-IC bypass surgeries, or (3) some with fewer than 10 previous cases received a provisional certification to perform EC-IC bypass on a participant enrolled in COSS under the supervision of the neurosurgical principal investigator or designate.

Both treatment groups were prohibited from undergoing any additional or subsequent surgical procedure that might alter cerebral hemodynamics or affect stroke risk, except carotid endarterectomy for development of symptomatic contralateral carotid stenosis.

For participants in the surgical group, preoperative and postoperative antithrombotic treatment was determined by the COSS neurosurgeon until they were returned to the antithrombotic treatment preferred by their physicians. Participants in the nonsurgical group continued to receive the antithrombotic treatment preferred by their physicians. Target goals for risk factor control were 130/85 mm Hg for blood pressure, 100 mg/dL for low density lipoprotein cholesterol, 150 mg/dL for triglycerides, and 7% for hemoglobin A1C.

Follow up and outcome measures

The first follow-up visit was 30–35 days after randomization. Surgical participants received a repeat PET scan 30–60 days postoperatively. Subsequent follow-up visits were at 3-month intervals until 24 months or the end of the trial. Each follow-up examination included (1) history and examination to identify new stroke performed by an investigator other than the operating neurosurgeon, (2) recording current medications by class, (3) National Institutes of Health Stroke Scale (NIHSS), modified Barthel Index, Rankin Scale, and Stroke Specific Quality of Life (SSQoL) assessment ^14–18, (4) monitoring of risk factors, and (5) Doppler examination to assess graft patency for the surgical group. Telephone follow-up was permitted if an in-person visit was impossible.

The primary endpoint for all participants randomized to the surgical group who received surgery was the combination of (1) all stroke and death from surgery through 30 days post-surgery and (2) ipsilateral ischemic stroke within 2 years of randomization. The primary endpoint in the nonsurgical group and for those randomized to the surgical group who did not receive surgery was the combination of (1) all stroke and death from randomization to randomization plus 30 days and (2) ipsilateral ischemic stroke within two years of randomization. Thus, those randomized to surgery who never underwent surgery were still analyzed in the surgical group but the 30 day period to count all stroke and death for the primary endpoint began at randomization, not at surgery. Ipsilateral ischemic stroke was defined as the clinical diagnosis of a focal neurological deficit due to cerebral ischemia clinically localizable within the territory of the symptomatic occluded internal carotid artery that lasted for more than 24 hours.

Secondary endpoints were all stroke, disabling stroke, fatal stroke, death, NIHSS, modified Barthel Index, Rankin Scale, and SS-QoL.^14–18 The combination of any stroke or death was added as a post hoc endpoint. All stroke was defined as the clinical diagnosis of a focal deficit due to ischemia or hemorrhage clinically localizable to the brain lasting more than 24 hours. Fatal stroke led directly to the participant’s death within 30 days of occurrence. Disabling stroke was defined as a modified Barthel Index of <12/20 at the first scheduled return visit more than 3 months after the stroke occurred.

All participants and their families were urged to contact the local study coordinator for any event that might be a stroke or in the event of death. The local site sent copies of all brain images and arteriography obtained for clinical purposes and any other relevant information to the statistical data and management center within 1 week. A summary that contained no information to identify treatment group was prepared for 2 members of the adjudication committee. If they disagreed, the summary was sent to a third adjudicator. If the third adjudicator did not agree with either of the first 2, there was a consensus vote among all 3. All stroke endpoints determined by the adjudication committee were classified into stroke subtypes according to the TOAST criteria.¹⁹

A local safety monitor reviewed monthly summary reports of all adverse events by blinded treatment assignment. NINDS appointed a DSMB that met at regular intervals.

Sample size calculation and data analysis

Sample size and power calculations assumed that the true primary outcome rates would be 40% in the nonsurgical group and 24% in the surgical group The eligibility criteria for COSS were selected based on our 1992–1997 prospective study of patients with symptomatic AICAO to match a high risk subgroup with hemispheric symptoms within 120 days and a specific OEF threshold who had an overall rate of ipsilateral stroke at 2 years of 40% on medical therapy.^3,20 Surgical morbidity and mortality were assumed to be 12% as in the EC-IC Bypass Trial and the 2-year post operative ipsilateral stroke rate was assumed be 12%, as it was for persons with normal OEF in our previous study.^3,5 For a 5% 2-sided test to have 90% power to detect this difference, 354 participants (177 per group) were required (nQuery Advisor Version 4, Statistical Solutions, Saugus, MA), To account for death from nonstroke causes, the sample size was increased by 5% to 372.

Baseline characteristics were compared using generalized Fisher exact tests for categorical variables and Wilcoxon Rank Sum tests for continuous variables. For the primary endpoint, rates were based on product limit estimates of 2-year rates and their standard errors. Participants were censored at their last follow-up visit. We did not impute values for participants who were not followed for the full 2 years. The test statistic was calculated as the difference in estimated rates divided by the standard error of that difference. The 2-sided z-statistic was compared to a standard unit normal distribution. All randomized participants were analyzed in the treatment group to which they were initially randomized (intention-to-treat principle). Secondary endpoints of any stroke, fatal stroke, disabling stroke, death and any stroke or death were analyzed using the same methods. The Barthel score was dichotomized as 19–20 versus ≤18, using a scale of 1–20. The Rankin score was dichotomized in 2 ways: 0–1 versus 2–6 and 0–2 versus 3–6. These are standard dichotomizations used in stroke clinical trials. ^21–23 For dichotomized outcomes, differences between treatments were compared using Fisher exact tests. Summary SS-QoL scores were compared using a t-test. An on-treatment analysis was performed by removing participants assigned to the surgical group who never underwent surgery and censoring on the day of surgery participants assigned to the nonsurgical group who underwent EC-IC bypass surgery.

The Lan and Demets method with the O’Brien-Fleming type spending function was used for interim analyses for efficacy assuming 2-sided 5% level test.²⁴ A futility analysis was performed at each interim analysis for efficacy. Thresholds for futility were computed in a manner similar to that used for interim stopping for efficacy with a newly developed method approved by the DSMB that controls the overall probability of a type II error at no more than 15% level, slightly more liberal than the 10% used for the primary efficacy outcome.²⁵ This method allows the trial to stop early for futility with larger conditional power than more conventionally used and subjectively determined fixed thresholds of 10–15%.²⁵ Conditional power was computed for 3 alternatives: for the null hypothesis, for the currently observed rates, and for the original design effect size. The conditional power applied to prespecified futility thresholds was that based on the assumption of the original design effect size. Futility analyses included only those participants who had completed the 2-year follow-up to allow for the expected early perioperative stroke rate in the surgical group to be counterbalanced by longer term reductions.

Early trial termination

The second interim analysis was conducted when 194 participants had been randomized. The futility analysis included 139 participants who had completed 2 year follow-up: 16 primary endpoints in 74 surgical group patients and 11 primary endpoints in 65 nonsurgical group patients. Conditional power based on the assumption of the original design effect size was 29%, which crossed the pre-specified futility boundary of 35%. The interim analysis for efficacy showed a between group difference in favor of the nonsurgical group of 5.0% (95% CI −12.5% to 11.5%). The DSMB considered redesigning the trial to detect a smaller absolute difference of 10% in favor of surgery. This would have required increasing the overall sample size from 372 to 986 to achieve 80% power. The DSMB recommended stopping the trial citing (1) the prespecified statistical boundary for declaring futility had been crossed using the design effect size and (2) given the unexpected relatively low rate of observed primary endpoints in the nonsurgical group, a clinically meaningful difference in favor of surgery would not be detectable without a substantial increase in sample size which was not feasible. The study was terminated on June 24, 2010.

In December 2010, the DSMB was notified of errors in the interim analysis program. One patient in the nonsurgical group with a vertebrobasilar stroke occurring after 30 days had been erroneously included in the efficacy analysis. Three additional primary endpoints in the nonsurgical group had not been included in the futility analysis. These had been adjudicated as primary endpoints by the adjudication committee but the computer program for the earlier analysis was not properly written with the correct event codes (2 were fatal strokes and 1 was an ipsilateral stroke occurring after a non-primary endpoint non-ipsilateral stroke). The conditional power recomputed based on the accurate data was 1% for the null hypothesis, 2% for the currently observed rates, and 50% for the original design effect size. The conditional power for the original design effect size did not cross the pre-specified futility boundary of 35%. Nevertheless, the trial remained closed. The analyses presented here based on a closure date of June 24, 2010 include all 195 participants randomized by that date and all endpoints as of that date, including the 3 not included in the original futility analysis.

RESULTS

Between June 2002 and June 2010, 195 participants were randomized: 97 to the surgical group and 98 to the nonsurgical group (Figure 1). Comparison of 24 baseline variables produced 1 difference at the p<.05 level in systolic blood pressure (Table 1). Follow-up for the primary endpoint until occurrence, 2 years, or the end of the trial was 99% complete (Figure 1). Median follow-up for the surgical group was 723 days (interquartile range [IQR], 288 – 730) and for the nonsurgical group was 722 days (IQR, 271 – 730). At last follow-up visit, risk factor control was similar in both groups (Table 2). Forty-three of the last available 3-month visits were by telephone, for which there were no measurements of risk factor control.

Figure 1.

Trial Flow Diagram. The most common reasons recorded for failure of screened participants to meet clinical eligibility criteria were transient ischemic attack or stroke not in territory of occluded carotid artery (1597), transient ischemic attack or stroke not within 120 days (485), language comprehension not intact (283), non-atherosclerotic condition causing carotid artery occlusion (280), modified Barthel index not ≥ 12/20 (168), unsuitable surgical candidate (160), no occlusion of 1 internal carotid artery (114), not competent to give informed consent (110), subsequent surgery planned that might alter cerebral hemodynamics (105), age not 18–85 (104). For each screened participant, only 1 reason needed to be provided.

Table 1.

Baseline Characteristics

	Surgical Group	Nonsurgical Group	P Valuea
Participants, No.	97	98
Age, mean(SD), years	58 (9)	58 (9)	.71
Male, No. (%)	69 (71)	61 (62)	.22
Race White, No. (%)	88 (91)	92 (94)	.65
Hypertension, No. (%)	76 (78)	77 (79)	.99
Atrial Fibrillation, No. (%)	1(1)	2(2)	.99
Hyperlipidemia, No. (%)	79 (81)	86 (88)	.45
Diabetes mellitus, No. (%)	21 (22)	23 (23)	.93
Cigarette smoking, No. (%)			.86
Current	33 (34)	37 (38)
Former	56 (58)	51 (52)
Never	7 (7)	9 (9)
Unknown	1 (1)	1 (1)
Previous Myocardial Infarction, No. (%)	11 (11)	14(14)	.83
Previous Stroke, No. (%)	44 (45)	35 (36)	.34
Entry Event Type, No. (%)			.39
Stroke	52 (54)	62 (63)
Transient Ischemic Attack	44 (45)	35 (36)
Unknown	1 (1)	1 (1)
Entry Event Side, No. (%)			.20
Right	54 (56)	45 (46)
Left	43 (44)	53 (54)
Entry Event to Randomization, mean(SD), days	72 (37)	75 (36)	.50
PET Ratio, mean(SD)	1.26 (0.14)	1.26 (0.14)	.76
Contralateral Carotid Stenosis (%)			.41
<50%	79 (81)	78 (81)
50% – 69%	11 (11)	7 (7)
≥ 70%	7 (7)	11 (11)
Unknown	0	2 (2)
Modified Barthel Scale (of 20), mean(SD)	19.3 (1.8)	19.5 (1.4)	.66
Modified Rankin Score, mean(SD)	1.4 (1.1)	1.4 (1.1)	.63
NIHSS, median(IQR)	1.0 (3.0)	1.0 (3.0)	.31
SSQoL (Summary,) mean(SD)	3.7 (0.9)	3.7 (0.8)	.78
Systolic Blood Pressure, mean(SD), mm Hg	133 (15)	139 (20)	.04
Diastolic Blood Pressure, mean(SD), mm Hg	76 (11)	77 (10)	.36
LDL-cholesterol, mean(SD), (mg/dL)*	107 (46)	105 (36)	.85
Triglycerides, mean(SD), (mg/dL)*	186 (118)	176 (167)	.17
Hemoglobin A1C, mean(SD), (%)*	6.0(1.0)	6.1 (1.1)	.29

Abbrevations: LDL, low density lipoprotein; NIHSS, National Institutes of Health stroke scale; PET, positron emission tomography; SSQoL, Stroke specific quality of life,.

^aP-values were computed using Fisher exact tests for categorical variables and Wilcoxon Rank Sum statistics for continuous variables.

Focal ischemic symptoms in the territory of the occluded carotid artery were categorized as cerebral transient ischemic attack (<24 hrs duration) or cerebral infarct (≥24 hours duration). LDL-cholesterol, triglyceride, and hemoglobin A1C values are from the 3-month follow-up visit.

The modified Barthel scale (0–20) evaluates the degree of independence in day-to-day self care activities; a higher score is better (indicates greater independence).The modified Rankin Score (0–6) is intended to define the degree of a participant’s functional disability; a lower score is better (indicates less functional disability).The NIHSS (0–42) is a quantitative neurological examination developed to measure the degree of neurological deficit due to stroke; a lower score is better (indicates less neurological deficit).The summary SS-QoL (1–4) asks how self-reported overall quality of life compares to before stroke; a higher score is better (indicates better quality of life).

Table 2.

Medical Therapy at Last Follow-up

	Surgical Group	Nonsurgical Group	P Valuea
Participants, No.	97	98
Use of anti-thrombotic medication, No. (%)	89/95 (94)	88/94 (94)	.99
Systolic Blood Pressure≤ 130 mm Hg, No. (%)	34/76 (45)	29/69 (42)	.87
Diastolic Blood Pressure≤ 85 mm Hg, No. (%)	62/76 (82)	48/69 (70)	.12
Cigarette smoking (not currently smoking) No. (%)	60/95 (63)	57/92 (62)	.79
LDL-cholesterol≤ 100 mg/dL, No. (%)	46/72 (64)	44/62 (71)	.46
Triglycerides≤ 150 mg/dL, No. (%)	45/75 (60)	42/64 (66)	.60
Hemoglobin A1C≤ 7%, No. (%)	65/74 (88)	53/64 (83)	.47

^aP-values from Fisher exact tests. LDL- low density lipoprotein

All primary endpoints were ipsilateral ischemic strokes, 20 in each group. In the surgical group, 19 were due to large artery atherosclerosis and 1 was undetermined. In the nonsurgical group, 16 were due to large artery atherosclerosis, 3 were undetermined, and 1 was due to small artery occlusion (lacune). For the intention-to-treat analysis of the primary endpoint, the 2-year rates were 21.0% (95% CI, 12.8%, 29.2%) for the surgical group and 22.7% (95% CI, 13.9%, 31.6%) for the nonsurgical group (p=0.78, z-test); difference = 1.7% (95% CI, −10.4%, 13.8%). (Table 3, Figure 2). The confidence interval of the difference excludes the original design effect size of 16% in favor of surgery. Given the data at termination, and if the true 2-year rate in the surgical group is 21.0%, then the true rate in the nonsurgical group would need to be more than 30% for the conditional power to exceed 20%. This is at the extreme of the 95% confidence interval of 31.6% for the rates at termination. Similarly, given the data at termination and if the true 2-year rate in the nonsurgical group is 22.7%, the true rate in remaining patients in the surgical group would need to be less than 14% for the conditional power to exceed 20%. This is less than the perioperative rate of 15% observed at termination (see below).

Table 3.

Intention-to-Treat Analysis

	Surgical Group (n=97)		Nonsurgical Group (n=98)
	Number of Events	Estimated Rate (95% CI)	Number of Events	Estimated Rate (95% CI)	Difference (95% CI)	P Value
Primary Endpoint	20	21.0 (12.8, 29.2)	20	22.7 (13.9, 31.6)	1.7 (−10.4, 13.8)	.78
Secondary Endpoints
Any Stroke	22	23.4 (14.8, 32.0)	24	26.9 (17.6, 36.2)	3.5 (−9.2, 16.1)	.59
Fatal Stroke	1	1.0 (0, 3.1)	2	2.4 (0, 5.6)	1.3 (−2.5, 5.2)	.50
Disabling Stroke	5	5.9 (0.8,10.4)	2	2.4 (0, 5.6)	−3.2 (−9.0, 2.6)	.27
Deaths	1	1.0 (0, 3.1)	5	5.1 (0.2, 9.9)	4.0 (−1.2, 9.7)	.13
Any Stroke or Death	22	23.4 (14.8, 32.0)	26	29.9 (20.1, 39.8)	6.5 (−6.5, 19.6)	.33
Modified Rankin score 0–1	48	49.5 (39.5, 59.4)	42	42.9 (33.1, 52.7)	−6.6 (−20.6, 7.3)	.41
Modified Rankin score 0–2	68	70.1 (61.0, 79.2)	73	74.5 (65.9, 83.1)	4.4 (−8.2, 16.9)	.70
Modified Barthel scale 19–20	68	70.1 (61.0, 79.2)	70	71.4 (62.5, 80.4)	1.3 (−11.4, 14.1)	.85
Summary SSQoL, mean		3.82 (3.60, 4.04)		3.58 (3.37, 3.79)	−.24 (−.54,. 07)	.13

The primary endpoint for all participants who were assigned to surgery and received surgery was the combination of: (1) all stroke and death from surgery through 30 days post surgery and (2) ipsilateral ischemic stroke within two years of randomization. The primary endpoint in the nonsurgical group and for those participants assigned to surgery who did not receive surgery was the combination of (1) all stroke and death from randomization to randomization plus 30 days and (2) ipsilateral ischemic stroke within two years of randomization. Ipsilateral ischemic stroke was defined as the clinical diagnosis of a focal neurological deficit due to cerebral ischemia clinically localizable within the territory of the symptomatic occluded internal carotid artery that lasted for more than 24 hours.

All stroke endpoints are two-year Kaplan-Meier estimates of the proportions (95% confidence intervals) based on 195 randomized participants. Deaths were of any cause within 2 years of randomization for all 195 randomized participants. There was 1 death (fatal stroke) in the surgical group on the day of surgery (day 7 after randomization). There were 5 deaths in nonsurgical group: 2 fatal strokes (days 67 and 657 after randomization) and 3 non-stroke deaths due to pneumonia (day 564), sudden death (day 597), and stage IV lymphoma (day 725).

Rankin and Barthel endpoints (proportions) and the SSQoL (stroke specific quality of life) have worst case imputed for deaths and missing values. All values are given with 95% confidence intervals. The p-values are for Fisher exact tests.

Figure 2.

Kaplan-Meier cumulative failure curves for the intention-to-treat analysis of the primary end point. The primary endpoint is defined in the footnote to Table 3.

The number of participants who remained event free and available for follow-up evaluation at each 90 day interval is shown in the appropriate color for each group at the bottom of the graph.

At 30 days, the rates of ipsilateral ischemic stroke were 14.3% (14/97) in the surgical group and 2.0% (2/98) in the nonsurgical group, difference = 12.4% (95% CI, 4.9%, 19.9%). For the intention-to-treat analyses of the secondary endpoints, there were no significant differences at the .05 level (Table 3).

Four participants randomized to the surgical group did not undergo surgery (Figure 1). One had an ipsilateral ischemic stroke (primary endpoint) 9 months after randomization. Three participants randomized to the nonsurgical group underwent ipsilateral EC-IC bypass (Figure 1). One had an early postoperative ipsilateral ischemic stroke (primary endpoint). For the on-treatment analysis, the 2-year estimates for the primary endpoint were 20.8% (95% CI, 12.4%, 29.1%) for the surgical group and 22.3% (95% CI, 13.3%, 31.2%) for the nonsurgical group (p=.81); difference = 2.0% (95% CI, −10.7%, 13.7%).

Other than crossovers, there were 3 surgical treatment protocol violations in the surgical group (1 ipsilateral common carotid stent, 2 asymptomatic carotid endarterectomies) and 1 in the nonsurgical group (contralateral asymptomatic carotid endarterectomy). One additional participant in the nonsurgical group underwent a contralateral carotid endarterectomy for symptomatic carotid stenosis as permitted by the protocol. Central arteriographic eligibility review of 191 studies (4 missing) revealed that 2 were not catheter arteriograms and 3 were not atherosclerotic occlusions.

Ninety-three participants in the surgical group underwent surgery a median of 6 days (IQR, 1 to 13) after randomization. No strokes occurred within this period. Within 30 days postoperatively, 14/93 (15%) experienced a stroke (all ipsilateral ischemic). Surgery was performed by 30 different surgeons. The surgical certification method was not significantly associated with 30 day postoperative stroke (chi-square Yates corrected .196, p= .91). Serious adverse events occurred in 12 additional participants within the 30 day postoperative period: 4 TIAs, 2 epidural/subdural hematomas, 2 seizures, 1 myocardial infarction, 1 respiratory disorder, 1 hypotension, and 1 wound infection. In the nonsurgical group, the only serious adverse events within 30 days of randomization were 2 primary endpoint ipsilateral ischemic strokes. Graft patency was 98% at 30 days (88/90 with data) and 96% at last-follow-up (86/90). The mean OEF ratio in the surgical group improved from 1.258 at baseline to 1.109 at the 30–60 day postoperative repeat PET scan (87 with data).

COMMENT

In spite of excellent graft patency and improved cerebral hemodynamics in the surgical group, EC-IC bypass surgery failed to provide an overall benefit on 2-year stroke recurrence. The 2-year primary endpoint rate of 21% in the surgical group was very close to the rate of 24% projected from historical data. The 30-day postoperative morbidity and mortality of 15% was not statistically significantly different from the rate of 12% (81/663) from the EC-IC Bypass Trial (p=0.44, Chi-square).⁵ After the 30-day postoperative period, the rate of recurrent stroke for the remainder of the 2-year follow-up period was 6%, half of the 12% rate projected. In contrast, the 2-year primary endpoint rate of 23% in the nonsurgical group was much lower than the rate of 40% projected from a prospective observational study of similar patients carried out from 1992–1997. The lower stroke risk observed in COSS for the nonsurgical group is similar to the better outcomes observed in more recent studies of patients with medically treated asymptomatic carotid artery stenosis, ascribed to improvements in medical therapy.²⁶ These observations reaffirm the hazard of using even the most carefully studied historical controls to infer therapeutic efficacy and the necessity of performing randomized controlled trials to establish clinical benefit. Although improved hemodynamics in participants who survived EC-IC bypass surgery without perioperative stroke was associated with low risk of recurrent stroke, the better-than-expected efficacy of medical therapy in the nonsurgical group was sufficient to nullify any overall benefit of surgery.

COSS was terminated based on a futility analysis that showed a 29% chance of demonstrating a statistically significant benefit for surgery if taken to completion, under the assumption that the original design effect size was true. The value of 29% was below the prespecified threshold of 35%. It was later determined that there was an error in the analysis and that the conditional probability under the assumption of the original design effect size was 50%. Computing conditional probability requires an assumption about the trends to be observed in the remainder of the study. Commonly used assumptions are the original design effect size (as used here for the threshold calculation) or the observed data.²⁷ Use of the original design effect size assumes that the reminder of the participants enrolled in the trial will come from a population with the endpoint rates as originally postulated (in this case 24% for the surgical group and 40% for the nonsurgical group). In COSS, the assumption of the original design effect size was reasonable for the surgical group. The perioperative stroke rate of 15% in COSS was essentially identical to the rate of 12% from the EC-IC Bypass Trial. The post-perioperative stoke rate in the COSS surgical group was 6% in two years (3% per year) as compared to 4% per year for the 385 participants with carotid occlusion in the surgical group of the EC-IC Bypass Trial.⁵ These data on the outcome of surgery are very consistent and it is unlikely that additional participants enrolled in the surgical arm of COSS would have substantially different overall outcomes. However, given the rate in the nonsurgical group of 22.7% with an upper 95% confidence bound of 31.6% observed at the time of the futility analysis, the assumption of the original design effect size of 40% for calculation of conditional power does not appear to be realistic. Using the alternative assumption based on the observed rates to calculate conditional power yields a probability of only 2% that the study would have rejected the null hypothesis if all 372 originally scheduled participants had finished 2-year follow-up. Thus, it is highly unlikely that COSS would have shown a statistically significant benefit for surgery if taken to completion.

Interpretation of the study is limited by the relatively small number of outcomes events. The 95% confidence bounds of the difference in the primary endpoint still allow for an absolute risk difference of 10% in favor of either group. Sham surgery was not performed, so there is the potential for bias in individual sites reporting potential endpoints for adjudication. This does not appear to have occurred given that the number of reported events adjudicated not to be events was 6 in the surgical group and 4 in the nonsurgical group.

The Japanese EC-IC Bypass Trial (JET) was similar in design to COSS. JET used the combination of reduced baseline cerebral blood flow and reduced cerebral blood flow increase in response to the vasodilator acetazolamide to identify patients with hemodynamic cerebral ischemia. From 1998–2002, 206 patients with major cerebral artery occlusive disease of the internal carotid artery or middle cerebral artery symptomatic within 3 months were enrolled. Final results of the 2-year follow-up were due in 2004. A second interim analysis of data from 196 patients followed through January 2002 reported primary endpoints in 14/98 medically treated patients and 5/98 surgically treated patients (p=.046 by Kaplan-Meier analysis). ^28,29 Examination of the published Kaplan-Meier curves show no endpoints within the first month in the surgical group. There is no explicit mention whether the results include 30 day postoperative morbidity and mortality, but it seems unlikely that this rate was 0 given that it was 12% in the original EC-IC bypass trial and 15% in COSS. We are not aware of publication of the final JET results.

COSS confirms the importance of hemodynamic factors in the pathogenesis of recurrent stroke in patients with symptomatic AICAO and the accuracy of PET measurements of OEF in identifying patients at high risk for recurrent stroke due to poor collateral circulation. The 23% rate of subsequent stroke at 2 years in these patients on medical therapy is comparable to that for patients with 70–99% symptomatic carotid stenosis.³⁰ Nevertheless, the results of COSS showed that EC-IC bypass surgery provided no additional benefit over medical therapy for preventing recurrent stroke.

COSS Personnel: Clinical Coordinating Center (University of North Carolina, Washington University): William J. Powers (Principal Investigator), Robert L. Grubb, Jr. (Neurosurgical Principal Investigator), Tom O Videen (PET Principal Investigator), Colin P. Derdeyn(Neuroradiology Radiology Principal Investigator), Carol Hess Papps, Karen McElvany, Lydia Counts, Claudia Kelly, Tori Barnard, Jessica Ferrari, Tena Davis, Tracy Hall. Statistical and Data Management Center (University of Iowa): William R. Clarke, Jr (Principal Investigator), Harold P. Adams, Jr. (Neurological Principal Investigator), Michele Costigan, Ken Singletary, Yi Fang, John Marmet, Dixie Ecklund, Michelle Wichman, Michael Hansen, Richard Peters, Diane Anderson, Robert F. Woolson, Cynthia Diltz, Julie Lang, Laura Beane, Jeri Sieren, Ying Zhang. Positron Emission Tomography Analysis Center (Washington University) Tom O. Videen (Principal Investigator), John T. Hood, Jr. Safety Monitors: Harold P. Adams, Jr., Bryce K. Weir. Endpoint Adjudication Committee: Thomas Brott, Philip Gorelick, Gary Ferguson. National Institutes of Neurological Diseases and Stroke: John Marler, Bernard Ravina, Wendy Galpern, Claudia Moy, Scott Janis. Data, Safety and Monitoring Board: E. Clarke Haley, Sean Grady, Yuko Palesch, David Eidelberg. Clinical Centers (in order of number enrolled):Washington University School of Medicine: Colin P. Derdeyn, Lina Shinawi, Nancy Hantler, Amber Tyler, Mary Catanzaro, Susie Fritsch, Ralph Dacey, Michael Chicoine, Gregory Zipfel. University of Iowa: Patricia Davis, Heena Olalde, Marge Rogers, Pam Jacobs, Patrick Hitchon, Enrique Leira, Matthew Howard, Jamal Hoballah, Jennifer Ireland. Columbia-Presbyterian Medical Center: Randolph Marshall, Kevin Slane, E. Sander Connolly, Joanne Festa Robert Solomon, Mandip Dhamoon, Joshua Wiley,, Guillermo Linares, Adrian Marchidann, ViktorSzeder, Olga Noskin, Dan Antoniello, Bernardo Liberato, Jennifer Frontera, Shyam Prabhakaran, J. Chong, Clinton Wright. University of Cincinnati: Mario Zuccarello, Carolyn Koenig, Lynn Sprafka, Matthew Flaherty, Brett Kissela, Dawn Kleindorfer,, Suzanne Kempinsky-Cliver, Joseph Broderick, Collkeen Roggee, Daniel Kanter, Daniel Woo, Alexander Schneider, Irene Ewing. University of Illinois Chicago: Fady Charbel, Kerriem Watson, Nadda Mlinarevich, Sean Ruland, Amin-Hanjani Sepideh, Rebecca Grysiewicz, Julie Jones. Wake-Forest University: John Wilson, Steven Glazier, Charlottte Miller, Wendy Jenkins, Akiva Mintz, Jeffrey Craig, Charles Tegeler, Kathleen O’Brien. Millard Fillmore Gates Hospital: Marilou Ching, Catalina Ionita, Peterkin Lee-Kwen, Richard Chan, Frederick Munschauer III, Kathleen Parkes, Mary Lou Coad, Kathleen Wrest, Karen Olson, Sharon Harrington, L. Nelson Hopkins, Walter Grand, Richard Ferguson, E. Marl Heckler, Lorraine Pereira, Robert Sawyer, James Budny, Eldad Levy, Lynn Nemeth, Jameela Ali. Cleveland Clinic: Peter Rasmussen, Doreen Andrews-Hinders, Irene Katzan, Ken Uchino, Marc Mayberg, Rishi Gupta, John Bartholomew, Jean-Luc Urbain, Terese Wheeler, Lori Strozniak. University of Arkansas: Ali Krisht, Sharon Partington, Deborah Fewell, James Schmidley. Stanford University: Gary Steinberg, Maria Coburn, Desiree Luu, Christine Wijman, Tim Darsaut, John Sinclair, Ralph Guzman, Anna Ciaravino, Monika Varga, Kathleen Biederman. Goodman Campbell Brain and Spine: Terry Horner, Troy Payner, Kathy Redelman,, James Fleck, John Scott. Emory University: Daniel Barrow, Emilee Holland, Fadi Nahab, Kathryn Rahimazadeh, Salina Waddy, Sally Burke, Marc Chimowitz, Michael Frankel, Owen Samuels, Sharion Sailor, Janey Braimah, Charlene Stark, Bethany Lane, Nicole Lessard, Harriet Howlett-Smith, Mustaffa Ezzedine, Barney Stern, Betty Shipp, Kate Hanson, Tanya Turan, Naema Osman. Indiana University: Christina Ivan, Askiel Bruno, Jose Biller, Flossy Nabila Lincoln, Sandy Guingrich, Angi Micheels, Michael Pritz, J.D. Fleck, Linda Williams,, William Jones, Alison Sears, Kelley Faber, Alfredo Lopez, Linda Chadwick, Vincent Matthews, Michael Girdler University of California, San Francisco: Michael Lawton, Lisa Hannegan, Wade Smith. University of Wisconsin: Robert Dempsey, Pam Winne, M.J. Washburn, Alice Li, Justin Sattin, Matthew Jensen, Marcus Chacon, Douglas Dulli, George Newman, Madlein Geraghty, Nicole Page, Jamie Kish. London Health Science Centre: Richard Chan, Kimberly Hesser, Steven Lownie, Vladimir Hachinski, Connie Frank, Lorraine Fleming, Karen Baptista, Suzan Matijevic. University of Pittsburgh: Tudor Jovin, Howard Yonas, Ken Uchino, Julia Billigen, Lori Kirby, Maxim Hammer, Mouhammad Jumaa, Ridwan Lin, Amer Malik, Nishant Ranawat, Vivek Reddy, Lawrence Wechsler, Syed Zaidi,, Nishant Ranawat, Mary Ann Rothfus,, David Adelson, Rishi Gupta, Ashis Tayal, James Gebel,, Edwin Nemoto, Ronda Pindzola, Yvette Mallory. University of Michigan Medical Center: B. Gregory Thompson, Donna Auer, Carol Persad, Dominique Powells, Darin Zahuranec, Devin Brown, Susan Hickenbottom. Thomas Jefferson University: Robert Rosenwasser, Ellen Simons, Maureen Jensen, Rodney Bell, Monisha Kumar, Michael Moussouttas, Carissa Pineda, Deborah August, Erol Veznedaroglu, DavidBrock, Gerri McGinnis, Rachel Scollon, Kathleen Hilbert. University of Texas, Southwestern Medical Center at Dallas: Mark Johnson, Hal Unwin, Dion Graybeal, Thomas Kopitnik, Lisa Davis, April Blair, Duke Samson, Jessica Lee,, Mounzer Kassab. University of California, Los Angeles: Neil Martin, Jeffrey Saver, Judith Guzy, Jill Haines, Jennifer Varma, Hannah Smith, Rana Fiaz, Latisha Ali, S. Thomas Carmichael, David Liebeskind, Bruce Obviagele, Paul Vespa, Kimberly Duke, Sid Starkman, Michael Froehler, Matthew Tenser, Pablo Villablanca, Chad Miller, Samer Shah, Johannes Czernin, Suichi Suzuki, Bruce Dobkin, Amytis Towfighi, Clelsea Kidwell, Andre Fredieu, Margaret Tremwel, Donna Manglona. St Joseph’s Hospital and Medical Center: Joseph Zabramski, JoAnn Snyder, James Frey, Robert Spetzler, Murray Flaster. Mercy General Hospital Sacramento: M. Asim Mahmood, Paul Akins, John Byer, John Schaefer, Deidre Wentworth, Susan Croopnick, Laura Newman, Kathleen Cabe. Park Nicollett Institute: Sandra Hanson, Amy Castle, Eric Nussbaum, Linda Kuno, Jennifer Mattson. University of New Mexico: Howard Yonas, Alice Brown, Marc Malkoff, Glen Graham, Elaine Stack, Lori Kirby. Wayne State University: Fernando Diaz, Richard Fessler, Vickie Gordon, Bradley Jacobs. Massachusetts General Hospital: Christopher Ogilvy, Deidre Buckley, Jaime Mansala, Allegra Bruce, Ferndinando Buonanno, David Greer, Meaghan Whalen. Mayo Clinic-Rochester: Frederic Meyer, Nicole Fode-Thomas, Steve Thalacker, David Piepgras, Jimmy Fulgham, John Atkinson. Loyola University Medical Center: Jose Biller, Linda Chadwick, Douglas Anderson, Michael Schneck, Rima Dafer, Sarkis Morales Vidal, Edward Duckworth. Case Western Reserve Medical Center: Warren Selman, Gerri Zalaski, Valerie Cwiklinski, Cathy Sila, Robert Tarr, Nicholas Bambakadis, Jose Suarez. Duke University Medical Center: Carmel Graffagnino, Joanna Stoner, Sarah Moore, Allan Friedman, Vani Chilikuri, Jessica Kesler. Baptist Memorial Hospital (Memphis, TN): Clarence Watridge, Mistina Pannell, Grace Miller Michael Jaciewicz, Lance Wright. Swedish Medical Center (Seattle, WA): David Newell, Beth Miller Kraybill, William Likosky, Brandi French, Jeannie Bush, Amanda Brown, Joanie Block, Bonnie Blanchard. University of Washington: Kyra Becker, Laligam Sekhar, DavidNewell, Patricia Tanzi, Dian Nuxoll, Heather Bybee, Steven Cramer. Methodist Hospital (Houston, TX): John Volpi, Marilyn Bautista, Jonathan Zhang, David Chiu, James Ling, Delmar Imperial-Aubin, Donald Bledsoe, Jose Tenorio. St. Louis University: Saleem Abdulrauf, Mandy Kane, Catharine Richard, Margaret Swoboda, Salvador Cruz-Flores, Brent Ibata. Providence Hospital (Southfield, MI): Richard Fessler, Kathryn Telck, Vickie Gordon Daniel Pieper, Bruce Silverman, Fernando Diaz, John Whapham. Healtheast Care System (Minneaplois, MN): Sandra Hanson, Amy Castle, Eric Nussbaum. University of Texas Health Sciences Center San Antonio: David Sherman, Anne Leonard, Laura Moreno, Silvina Tonarelli, Robert Hart, Merrill Carolin, Gustavo Roman, Jean-Louis Caron, Oscar Benevente, Dennis Vollmer. Boone Hospital Center (Columbia, MO): Allyn Sher, Stacey Jett, Gerry Neely. West Los Angeles Veterans’ Administration Medical Center: Roi Ann Wallis, Kim Panizzon, Eliott Licht, Eric Cheng. St. Francis Medical Center (Cape Giradeau, MO): Robert Gardner, Janey McAdoo, Darlene Haupt, Kathleen Riggs-Ruupp. Michigan State University: Mounzer Kassab, Susan Wehner, Arshad Majid. Ohio State University: Andrew Slivka, Julie Agriesti. University of Kentucky: Michael Dobbs, Anna Fowler, L. Creed Pettigrew, Anand Vaishnar, Stephen Ryan, Sherry Williams. Medical College of Georgia: Fenwick Nichols, Brian Close, Askiel Bruno, Jeffrey Switzer, David Hess, Christiana Hall, Joanne Rogalsky-Nacca. University of North Carolina at Chapel Hill: Susan Wilson, Natalie-Aucutt-Walter, David Huang, Roxanne Poole, Matthew Ostrander. St. John’s Mercy Medical Center (St. Louis, MO): William Logan, Sally Schroer. MainlIne Health Stroke Program: Gary Friday, Joan Brown. PET Centers (in order of number of scans performed): Washington University School of Medicine: Colin P. Derdeyn, Lennis Lich, Sally Schwarz, Greg Gaehle. Kettering Medical Center (Kettering, OH): Joseph Mantil, Pamela Strohmeyer, Elaine Gagermeier, Steve Mattmuller, T.K. Narayanan, Binzhi Shi, MartinSatter, Gordon Simmons, Brad Christian, Max Carone, Tonya Perkins, Marilyn Brackney. University of Pittsburgh: James Mountz, Julie Price, Donald Sashin, Carolyn Meltzer, Chet Mathis. Northern California Positron Imaging Center: Elma Abella, Dan Cariddi, Bruce Finley, Peter Valk, Jan Cronin, Ruth Tesar. University of Iowa: Richard Hichwa, John Sunderland, Len Watkins, Laurie Ponto. Columbia-Presbyterian Medical Center: Ronald Van Heertum, Norman Simpson, Chitra Saxena. SUNY Buffalo/Western New York VA: Alan Lockwood, Michael Haka. Wayne State University: Harry Chugani, Lynda Ferguson, Otto Musik, Tom Manger. Good Samaritan Medical Center Phoenix, AZ) : Eric Reiman, Debbie Intorcia, Stephanie Reeder, Dan Bandy, Tricia Giurlani, Anita Prouty, Christine Burns. University of Wisconsin: Charles Stone, Brad Christian, Alex Converse, Andrew Roberts, Terry Oakes, Kevin Dabbs. Mayo Clinic-Rochester: Brian Mullan, Mark Jacobson, Brad Kemp, Terry Brinkman. Wake-Forest University: Pradeep Garg, Kathryn Morton, Paul Stroud, Kerry Link, Clive Brown-Proctor, Fred Fahey, Scott Wollenweber, Holly Smith. University of Texas Health Science Center (San Antonio): Peter Fox, Jean Hardies, Laura Morena, Shalaini Narayana, Paul Jerabek. University of California, Los Angeles: John Mazziotta, Roger Woods, Olga Fernandez, Nazim Jaffer. Massachusetts General Hospital: Tom Brady, Alan Fischman, Ron Callahan, Jack Correi. Emory University: Mark Goodman, Ron Crowe, John Votaw, J. Douglas Bremner, Jonathan Nye. University of Washington: Janet Eary, Kenneth Krohn, Kristin Kauno, Barbara Lewellen, Linda Kimura, Steve Shoner, Tom Lewellen. Duke University Medical Center: R. Edward Coleman, Neil Petry.