Affiliations of authors: Department of Paediatrics, Division of Haematology/Oncology, The Hospital for Sick Children (PCN, LS), Department of Medicine, Princess Margaret Hospital, University Health Network (MC), Department of Health Policy Management and Evaluation (LS), Department of Public Health Sciences (JB), The University of Toronto, Toronto, Ontario, Canada; Program in Population Health Sciences, The Hospital for Sick Children, Toronto (LS, JB).
Correspondence to: Lillian Sung, MD, FRCPC, The Hospital for Sick Children, Division of Haematology/Oncology, 555 University Ave., Toronto, Ontario, M5G 1X8, Canada (e-mail: lillian.sung{at}sickkids.ca)
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ABSTRACT |
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INTRODUCTION |
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We performed a meta-analysis of all relevant randomized trials that compared autologous bone marrow transplantation with further non-myeloablative chemotherapy in adult patients with acute myeloid leukemia. Because the majority of these patients do not have a histocompatible donor, most of these trials used a method of biologic randomization in which all eligible patients with a matched sibling donor are allocated to an allogeneic transplantation arm, and all remaining patients are randomly assigned to receive either autologous bone marrow transplantation or chemotherapy (4).
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METHODS |
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Studies were eligible for inclusion in the meta-analysis if they conformed to one of two trial designs: 1) prospective cohort studies that offered an allogeneic bone marrow transplant to all eligible patients in first remission with an available sibling donor (biologic randomization) and randomly assigned all remaining patients to either autologous bone marrow transplantation or chemotherapy (or no further treatment) and 2) randomized trials that compared autologous bone marrow transplantation with chemotherapy (or no further treatment) in all patients in first remission. Trials conforming to one of these two designs were eligible if their samples were restricted to patients aged 1555 years or if information regarding participants in this age range could be extracted from the publication. Studies that focused exclusively on children or on patients older than 55 years were excluded. Although the trials included in our study were required to include patients with primary or de novo acute myeloid leukemia, trials that included patients with secondary acute myeloid leukemia were also eligible as long as those patients were not the exclusive focus of the trial.
Search Strategy
Potentially relevant studies in all languages were initially identified by a comprehensive search strategy that included an electronic search of MEDLINE/PreMEDLINE (January 1966 through September 2002), EMBASE (January 1980 through September 2002), the Cochrane Controlled Trials Registry (through the third quarter of 2002), and CancerLit (January 1975 through August 2002) as well as a manual search of the reference lists of articles and pertinent reviews that were retrieved in full. Two reviewers (P. C. Nathan and L. Sung) independently evaluated the titles and abstracts of all publications identified in the initial literature search, and any publication considered to be potentially relevant by either reviewer was retrieved in full for further assessment. The reviewers were not blinded to the authors of the study or study outcomes. Of the publications retrieved for further assessment, only those considered eligible by both reviewers were included in the final meta-analysis. Disagreements were resolved by consensus; in the absence of consensus, the opinion of a third reviewer (M. Crump) was sought. A kappa score was calculated for the inter-rater agreement on which publications should be included in this meta-analysis. The strength of inter-rater agreement, as evaluated by the kappa score, was defined as slight (0.000.20), fair (0.210.40), moderate (0.410.60), substantial (0.610.80), or almost perfect (0.811.00), as previously described (9).
Validity Assessment
Publications included in the final meta-analysis were independently assessed for validity by the original two reviewers with the use of an eight-item checklist that assessed whether all eligible patients were tested for the presence of a sibling donor, all patients with a sibling donor were assigned to the allogeneic bone marrow transplantation arm, all remaining patients were randomly assigned to autologous bone marrow transplantation or continued chemotherapy (or no further treatment), randomization was centralized, allocation was concealed, randomization was stratified, outcome assessors were blinded, and outcomes were assessed using intent-to-treat analysis. In addition, completeness of follow-up was evaluated on a four-point ordinal scale (<1%; 1%5%; >5%20%; >20%) that was based on our consultations with colleagues who have expertise in the field. A kappa score for inter-rater agreement on study validity was calculated, and its strength was defined as described above.
Data Abstraction
Data were independently abstracted by both reviewers with the use of a standardized data abstraction form, and consensus was reached on any disagreement. Study authors were contacted if important data were not available in the published study.
Statistical Analysis
The primary outcomes in our study were overall survival and disease-free survival. For most of the eligible studies, we could not estimate hazard ratios for survival because we could not obtain the necessary information from the published manuscript or by contacting the authors. Therefore, we derived a point estimate of the ratio of survival probabilities at 48 months comparing the autologous bone marrow transplantation group with the chemotherapy group by using the KaplanMeier estimates of survival (and their standard errors) as reported in the studies. The variance of each ratio was determined by using the delta method (10). When 48-month survival data could not be determined from information provided in the publication or by contacting the authors, we used data from the closest available time point. Effect sizes were weighted on the basis of their inverse variance. Study heterogeneity was assessed with the Cochrane Q test (11) using a predefined statistical significance level of .10. It was decided, a priori, that data would be pooled using a fixed-effects model if there was no evidence of heterogeneity between studies.
Secondary outcomes of interest included death during remission (i.e., treatment-related mortality) and the proportion of patients who achieved and remained in a second remission if they relapsed after consolidation therapy (i.e., salvage rate). Categorical data were compared using a chi-square test. Publication bias was assessed by visual inspection of a funnel plot of the effect size (i.e., ratio of disease-free survival probabilities) versus its precision (i.e., the inverse of its variance) for each of the included studies. We assumed that asymmetry in the lower left-hand corner of the funnel plot indicated that small trials with negative outcomes were not represented in the published literature (i.e., publication bias). The potential effect of such missing trials was evaluated using the trim-and-fill technique (12). All analyses were based on the treatment groups to which patients were originally assigned (i.e., intent-to-treat analysis). Statistical and graphical analyses were performed using Review Manager, version 4.2.2 (The Cochrane Collaboration, Oxford, U.K.), Microsoft Excel 2002 (Microsoft, Redmond, WA), and SPSS, version 11.0.1 (SPSS, Chicago, IL). All statistical tests were two-sided.
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RESULTS |
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The electronic database search yielded 574 potentially relevant publications, and a manual search of their reference lists and those of 22 review articles identified an additional 13 potentially relevant publications. The titles and abstracts of all 587 publications were examined by both reviewers, and 36 publications were retrieved in full for further consideration. Of the 36 publications reviewed in full, six (1318) reported studies that satisfied the inclusion criteria. Of the 30 publications excluded from this meta-analysis, 12 were either interim reports or abstracts of studies that were reported in full elsewhere (2,1929) and 18 reported studies that did not contain a chemotherapy arm or did not prospectively randomly assign patients to either autologous bone marrow transplantation or chemotherapy (1,3,3045). The kappa statistic for inter-rater agreement on the selection of the included studies was 1.00 (P<.001), indicating perfect agreement.
In all of the included studies, all eligible patients in first remission were tested for the presence of a matched sibling donor. Patients who did not have a matched sibling donor were randomly assigned to receive autologous bone marrow transplantation or chemotherapy (1317) or to receive autologous bone marrow transplantation or no further treatment (18). Table 1 lists the characteristics of the studies included in this meta-analysis, including the eligibility criteria used for patient randomization and interventions.
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Sample Size
A total of 4058 patients received induction chemotherapy in the six studies (Table 2). Of these, 2989 (74%) patients achieved a first remission (range = 65%80%). After exclusion of patients who had died, relapsed, been allocated to allogeneic bone marrow transplant, or withdrawn from the study for other reasons, 1044 (26%) patients (range = 18%41%) were randomly assigned to receive either autologous bone marrow transplantation or chemotherapy (or no further treatment). Of the 524 patients randomly assigned to receive autologous bone marrow transplantation, 360 (69%) underwent their intended transplant (range = 54%87%), whereas 480 (92%) of the 520 patients randomly assigned to receive chemotherapy (or no further treatment) received their assigned therapy (range = 83%100%) (odds ratio [OR] = 0.18, 95% confidence interval [CI] = 0.13 to 0.27; P<.001). Among the 164 patients who did not receive the autologous bone marrow transplant to which they were assigned, reasons for noncompliance were reported for 106 patients. These included relapse (39 patients [37%]), refusal (36 patients [34%]), toxicity (13 patients [12%]), and insufficient quantity of stem cells harvested (10 patients [9%]). Among the 40 patients who did not receive the non-myeloablative chemotherapy to which they were assigned, reasons for noncompliance included refusal (15 patients [38%]), relapse (10 patients [25%]), and toxicity (eight patients [20%]).
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We calculated overall survival and disease-free survival at 48 months for four of the six studies (1518) (Table 3). The fifth study presented data for overall survival and disease-free survival at 36 months (14), and the sixth study presented data for disease-free survival at 30 months only (13). We could not obtain 48-month survival data for the latter two studies. However, because the survival curves presented in those studies were relatively stable after 30 months (13) and 36 months (14), we used the survival estimates at those time points to calculate the pooled ratio of survival probabilities. These two studies accounted for 1.32% (13) and 9.38% (14) of the total weight of the six studies, suggesting that their impact on the pooled estimate of survival was minimal.
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Patients randomly assigned to receive autologous bone marrow transplantation had a statistically significantly greater risk of death during first remission than patients randomly assigned to receive chemotherapy or no further treatment. There were 57 (10.9%) deaths during first remission among the 524 patients who received autologous bone marrow transplantation and 23 (4.4%) deaths during first remission among the 520 patients who received chemotherapy or no further treatment (OR = 2.63, 95% CI = 1.60 to 4.32; P<.001) (Fig. 3). This odds ratio represents a 6.45% (95% CI = 3.26% to 9.65%; P<.001) difference in the risk of death in first remission between the treatment groups. The increased risk of death during first remission for patients receiving autologous bone marrow transplantation remained statistically significant, even when we excluded data from the (UK) Medical Research Council Acute Myeloid Leukemia 10 (MRC AML 10) trial (18), in which patients were randomly assigned to a no-further-treatment arm rather than a chemotherapy arm (OR = 2.33, 95% CI = 1.28 to 4.25; P = .006).
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Among the 524 patients randomly assigned to receive autologous bone marrow transplantation, 232 patients (44%; range = 35%60%) relapsed compared with 303 (58%; range = 53%80%) of the 520 patients randomly assigned to receive chemotherapy or no further treatment (P<.001) (data not shown). Because of reporting differences between individual studies, we could not generate a pooled estimate of the proportion of patients who achieved and remained in second remission after their first relapse. In four studies (1315,17), 14% (range = 9%22%) of patients who relapsed after receiving chemotherapy had achieved a second remission and remained in remission when the original studies were published. Three of those studies (13,15,17) reported that 5% (range = 0%11%) of patients randomly assigned to the autologous bone marrow transplantation arm remained in second remission at the time of publication of the original study. Furthermore, in the MRC AML 10 trial, the proportion of relapsed patients who achieved a second remission was statistically significantly higher in the no-further-treatment arm than in the autologous bone marrow transplantation arm, although this difference did not translate into a better 2-year survival after relapse (18).
Publication Bias
A funnel plot of the ratio of disease-free survival probabilities versus the inverse of their variances was asymmetric because of the large positive result reported in the BGM 84 trial (13) and suggested that the results of small negative trial(s) might not have been published (data not shown). To assess the impact of this potential publication bias on our point estimate of disease-free survival, we performed a second analysis that excluded the results of the BGM 84 study. The pooled fixed estimate of disease-free survival for the five remaining studies was 1.22 (95% CI = 1.05 to 1.43) (data not shown). We then added back a hypothetical study with a negative effect and weighting that were equivalent to the positive effect and weighting of the BGM 84 study. Recalculation of disease-free survival for the six original studies plus the hypothetical study produced a ratio of survival probabilities (1.22, 95% CI = 1.05 to 1.42) that was similar to that (1.24) for the original six studies. Therefore, we conclude that the possible exclusion of a small negative trial from this analysis would have minimal impact on our results.
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DISCUSSION |
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Two factors may account for our finding that improved disease-free survival did not translate into improved overall survival. First, 14% of the patients who relapsed after receiving non-myeloablative chemotherapy achieved a sustained second remission after undergoing an allogeneic or autologous bone marrow transplant. Second, patients who received autologous bone marrow transplantation had a 6.45% greater risk of dying during remission than patients who received non-myeloablative chemotherapy or no further treatment, primarily because of the toxicity associated with the transplant conditioning regimen. For example, patients who undergo autologous bone marrow transplantation for acute myeloid leukemia often experience protracted pancytopenia, which results in a prolonged risk of infection or bleeding (32). Hematopoietic recovery can be hastened by the use of autologous peripheral blood stem cells instead of bone marrow for the transplant (46). If a reduction in treatment-related mortality is associated with the use of autologous peripheral blood stem cells, patients who receive transplantation might experience improved overall survival compared with those who receive continued chemotherapy. However, a considerable reduction in treatment-related mortality would be needed to demonstrate a survival benefit from autologous peripheral blood stem cell transplantation.
Although allogeneic transplantation was once considered the preferred consolidation therapy for all young patients with acute myeloid leukemia, recent evidence suggests that this approach should be reserved for patients at the greatest risk of relapse, such as those with unfavorable cytogenetic features or a poor response to induction therapy (47). Greater uncertainty exists regarding how acute myeloid leukemia patients in first remission should be stratified by risk for autologous bone marrow transplantation (48). Resolution of this question has proven difficult, in part because of small sample sizes and disparities in the classification of risk in different studies. Insufficient information was provided in the publications included in this meta-analysis to assess the benefits of risk stratification. A meta-analysis that combines the individual patient data from each study (rather than just combining the study results) would be useful in addressing this question and would also give the most precise estimate of pooled time-to-event data (49).
The overall quality of a meta-analysis is contingent on the quality of the included trials (50). Several methodologic issues that are common to most trials of bone marrow transplantation must be considered when interpreting the results of this meta-analysis. For example, patients in such trials are randomly assigned to receive allogeneic transplantation or alternative consolidation modalities based on the availability of a matched sibling donor. This process of biologic randomization is generally considered to be an acceptable methodology when true randomization is not feasible, because it assumes that there are no systematic differences between patients who do and do not have a histocompatible donor (49). It has been suggested that the availability of a sibling donor may lead to confounding among acute myeloid leukemia trials in children, because the age of the patient may be related to the number of potential sibling donors as well as to the patient's chance of cure (51). Such confounding would not be a source of bias among acute myeloid leukemia trials in adults.
A greater source of bias is the large number of patients who are eligible for being randomly assigned to autologous bone marrow transplantation or an alternative therapy but are not, often because of physician or patient concerns about treatment-related toxicity, poor hematologic recovery, or uncertainty about the benefit of randomization (18). Furthermore, many patients who are randomly assigned to one treatment or the other do not receive their intended therapy (32). In our pooled analysis, we found that only 69% of patients randomly assigned to autologous bone marrow transplantation received transplants during first remission, whereas 92% of patients randomly assigned to chemotherapy or no further treatment received their assigned treatments. Although the use of intent-to-treat analyses has been advocated to minimize the bias associated with poor treatment compliance (8), such an approach may underestimate the true efficacy of transplantation (52). The timing of randomization can also have an impact on the assessment of treatment efficacy. Delaying randomization until late in therapy selects for patients who have a better chance of survival, leading to a time-lag bias (8). Conversely, early randomization may lead to an underestimation of efficacy because many patients will die or relapse before they receive their intended therapy. In this regard, it has been suggested that a common standard for measuring survival time, such as measuring from the day of human leukocyte antigen (HLA) typing, should be adopted to minimize reporting differences between studies (53).
Our study has several limitations. First, our use of a funnel plot to detect publication bias was limited by the small number of available studies. Our analysis, however, did demonstrate an important strength of meta-analysis: the enhanced power to detect a statistically significant difference between treatment arms when combining the results of similar studies. Although only one of the six studies included in our analysis reported a statistically significant disease-free survival advantage for the autologous bone marrow transplantation group, the combined data from all six studies resulted in a statistically significant pooled estimate of disease-free survival associated with autologous bone marrow transplantation.
A second limitation of our study was the lack of sufficient information available from the published studies or by contacting the study authors to allow us to calculate, either directly or indirectly, the log hazard ratio and its variance, which are the preferred summary statistics for reporting time-to-event data (54). As an alternative, we used survival at a fixed time-point, 48 months, as our endpoint. This approach is reasonable because the majority of events in patients with acute myeloid leukemia occur within the first 3 years after diagnosis (49). In the BGM 84 (13) and BGMT 87 trials (14), survival data were available only at 30 months and 36 months, respectively. However, recalculation of the pooled estimates of overall survival and disease-free survival with these trials excluded did not alter the point estimates for survival.
The role of autologous bone marrow transplantation in acute myeloid leukemia remains a topic of ongoing disagreement between the large cooperative study groups (8). Our results suggest that autologous bone marrow transplantation in first remission does not improve overall survival in patients with acute myeloid leukemia. Our results do not support the use of autologous bone marrow transplantation in all eligible patients; instead, they support the use of non-myeloablative chemotherapy during first remission for patients who do not have a matched sibling donor, with the option of salvage therapy with autologous bone marrow transplantation or transplantation from an unrelated donor for those who relapse. Further study is needed to determine if specific subgroups of patients in first remission, such as those with cytogenetic features associated with poor risk, might benefit from autologous bone marrow transplantation. Alternatively, this question could be addressed by combining individual patient data from the completed trials, but such an endeavor would require a large investment of resources as well as multinational cooperation.
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NOTES |
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We are grateful to Dr. Mark Greenberg for his insightful comments on an earlier draft of the manuscript.
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Manuscript received April 22, 2003; revised October 30, 2003; accepted November 7, 2003.
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