Patient Survival after Human Albumin Administration

Albumin administration has played a part in fluid management of acutely ill patients for more than 50 years (1). Indications for which albumin therapy is licensed in the United States include hypovolemia or shock; burns; hypoalbuminemia or hypoproteinemia; surgery; trauma; cardiopulmonary bypass; the acute respiratory distress syndrome; hemodialysis; acute nephrosis; hyperbilirubinemia; acute liver failure; ascites; and sequestration of protein-rich fluids in acute peritonitis, pancreatitis, mediastinitis, and extensive cellulitis (2-6). Albumin therapy has multiple effects, including volume expansion (7, 8), increased serum albumin concentration (9) and colloid osmotic pressure (10), and hemodilution (8). Albumin may be used expressly to achieve one or more of these effects. For instance, in high-risk neonates, albumin therapy may be given to expand intravascular volume and correct hypoalbuminemia (11). Both volume expansion and hemodilution may be desired in cardiopulmonary bypass surgery (12, 13) and acute ischemic stroke (14). A previous meta-analysis of randomized, controlled trials examined the effect of albumin on mortality compared with crystalloid, no albumin, or a lower dose of albumin (15). No significant effect of albumin on mortality was observed in trials of hypovolemia from surgery or trauma. However, in trials of burns and hypoalbuminemia, the mortality rate in albumin recipients significantly exceeded that in controls. When results of all three indications were pooled, the mortality rate was also significantly higher in the albumin group. The meta-analysis has been updated (16), but no further trials with mortality data were included and the results remained unchanged. The validity of the previous meta-analysis has been challenged on several grounds, such as omission of relevant trials, combination of heterogeneous trials (for example, of adults and high-risk neonates), inadequate assessment of the effect of methodologic quality on outcome, and the absence of a plausible mechanism to explain albumin-associated excess mortality (17-23). Independent evidence also suggests that the conclusion of the meta-analysis may have been incorrect. First, despite widespread use over an extended period, very few fatal adverse events have been reported in connection with albumin administration (1, 24-28). In addition, in a randomized, controlled trial of cirrhotic patients with spontaneous bacterial peritonitis, albumin significantly reduced mortality and renal impairment (29). This evidence suggests that albumin administration is not associated with excess mortality. We therefore conducted a new meta-analysis of randomized, controlled trials to test this hypothesis. Methods Trial Inclusion Criteria We used broad inclusion criteria and planned analyses a priori to investigate the relationship between methodologic quality and outcome. This approach avoids the potential for particular exclusion rules to bias meta-analytic results (30). Broad meta-analyses also offer the advantages of statistical power and generalizability (31). To be included, a trial had to be a randomized comparison of albumin therapy with crystalloid therapy, no therapy with exogenous purified albumin, or a lower dose of exogenous purified albumin; in addition, mortality data had to be available. We placed no restrictions on the clinical indication for albumin administration or therapeutic intent, such as volume expansion, correction of hypoalbuminemia, hemodilution, or maintenance of colloid osmotic pressure. Regimens in the control group could include crystalloid solutions but not synthetic colloids, blood products, or plasma protein fraction. We did not specifically exclude studies in which concomitant treatments, including asanguinous colloids and blood products, were given in a similar manner to both study groups. Search Techniques We sought published and unpublished trials through computer searches of the MEDLINE and EMBASE databases, the Cochrane Controlled Trials Register, and the Cochrane Medical Editors Trial Amnesty of unpublished trials. Relevant Internet resources, such as conference reports, abstracts, compilations of references, and full-text journal articles, were accessed by using the Altavista, Northern Light, HotBot, and Excite search engines. We manually searched Journal of the American Medical Association, The New England Journal of Medicine, The Lancet, and BMJ for the period January 1990 to November 2000. The medical directors of albumin suppliers and authors of published randomized trials related to albumin were consulted. The reference citations in previous meta-analyses, including those in the Cochrane Database of Systematic Reviews (both completed reviews and protocols), review articles, and reports of randomized and other investigations involving albumin were examined. Language restrictions were not applied. Data Collection We selected trials and independently extracted data. Differences in interpretation were resolved by discussion. Interanalyst agreement in trial selection and data extraction was not quantified. Authorship, time periods of enrollment, treatment protocol details, patient demographic data, and other information in the randomized trial reports were closely examined to avoid repeated inclusion of data on patients who may have appeared in multiple reports and to ensure inclusion of the most thorough data. During trial selection and data extraction, we were not masked to authors, institutions, journal, or identity of treatment groups. Available evidence indicates that for purposes of meta-analysis, such masking imposes substantial burdens without significantly altering effect sizes (32). Masking would also have hindered efforts to avoid duplication and ensure completeness of data. Deaths were scored on an intention-to-treat basis. The method used to conceal randomized group allocation was classified as adequate, inadequate, or unclear by using criteria described elsewhere (33). Unpublished data on mortality were sought from the randomized trial investigators. Statistical Analysis Our primary outcome measure was relative risk for death. Relative risks less than 1.0 or greater than 1.0 signify lower risk in the albumin or control group, respectively. We used the METAN program (34) to calculate relative risk and corresponding 95% CIs according to the method of Mantel and Haenszel (35) and to assess between-study heterogeneity according to the method of DerSimonian and Laird (36). If significant heterogeneity was not found, a fixed-effects model was used to calculate pooled relative risk and 95% CIs. Sensitivity analyses were planned and conducted to investigate the effects of the following methodologic quality attributes on outcome: blinding, allocation concealment method, presence of mortality as a study end point, and crossover. We used the METABIAS program (37) to evaluate small-trial bias according to the test proposed by Egger and colleagues (38). Frequently termed publication bias, small-trial bias reflects a systematic difference in effect size estimates derived from small versus large trials. Several mechanisms, including selective publication of small trials with positive results, bias in identifying relevant small trials, true heterogeneity between small and large trials, and poor quality of small trials, can produce publication bias (38). Role of the Funding Source This study was conceived and initiated by the two investigators. Support was thereafter sought and received from the Plasma Protein Therapeutics Association and the American Red Cross. These organizations had no role in the design, conduct, interpretation, and analysis of the study or in review and approval of the manuscript. Results Included Trials Figure 1 shows the number of trials evaluated during each stage of the meta-analysis. After initial screening and subsequent more detailed evaluation of retrieved randomized trial reports, 78 candidate trials including a total of 4232 randomly assigned patients were identified (7-14, 29, 39-108). Data on mortality were available from 55 trials involving 3504 randomly assigned patients; these trials accounted for 83% (3504 of 4232) of the randomly assigned patients in all candidate trials. These 55 trials, in which 525 deaths occurred, were included in the meta-analysis. No study was unpublished. Mortality data appeared in the published reports of 45 trials; the investigators provided unpublished mortality data for 10 trials. Despite our best efforts, we were unable to obtain mortality data for 23 trials. For 1 trial, details of the investigational design appeared in a published abstract (103) and mortality data appeared in a subsequently published letter (104). Figure 1. Number of trials evaluated at each stage of the meta-analysis. Of the 55 included trials, 27 involved surgery or trauma; 4 involved burns; 5 involved hypoalbuminemia; 6 involved high-risk neonates; 5 involved ascites; and 8 involved other indications, comprising 2 trials each of the acute respiratory distress syndrome and hyperbilirubinemia and 1 trial each of septic and hypovolemic shock, acute ischemic stroke, the vascular leak syndrome, and the ovarian hyperstimulation syndrome (Table 1). The median number of patients who underwent randomization per trial was 52 (range, 10 to 300). In addition to the 6 trials of high-risk neonates, 2 trials in the other indications category involved infants with hyperbilirubinemia (Table 1). Forty-five trials involved adults only. Age was unspecified in 2 trials. Table 1. Attributes of Included Trials Duration of follow-up either was specified or could be imputed in 49 trials. In these trials, the median duration of follow-up was 11 days (range, 0.04 to 1096 days). In 1 trial, patients were followed until discharge or death (9). Duration of follow-up was unspecified for 5 trials. Regimens Fluid management regimens and concomitant therapies varied widely (Table 1). In 7 trials, in add