We sought to examine the relation between sodium bicarbonate prophylaxis for contrast-associated nephropathy (CAN) and mortality. We conducted an individual patient data meta-analysis from multiple randomized controlled trials. We obtained individual patient data sets for 7 of 10 eligible trials (2,292 of 2,764 participants). For the remaining 3 trials, time-to-event data were imputed based on follow-up periods described in their original reports. We included all trials that compared periprocedural intravenous sodium bicarbonate to periprocedural intravenous sodium chloride in patients undergoing coronary angiography or other intra-arterial interventions. Included trials were determined by consensus according to predefined eligibility criteria. The primary outcome was all-cause mortality hazard, defined as time from randomization to death. In 10 trials with a total of 2,764 participants, sodium bicarbonate was associated with lower mortality hazard than sodium chloride at 1 year (hazard ratio 0.61, 95% confidence interval [CI] 0.41 to 0.89, p = 0.011). Although periprocedural sodium bicarbonate was associated with a reduction in the incidence of CAN (relative risk 0.75, 95% CI 0.62 to 0.91, p = 0.003), there exists a statistically significant interaction between the effect on mortality and the occurrence of CAN (hazard ratio 5.65, 95% CI 3.58 to 8.92, p <0.001) for up to 1-year mortality. Periprocedural intravenous sodium bicarbonate seems to be associated with a reduction in long-term mortality in patients undergoing coronary angiography or other intra-arterial interventions.
Contrast-associated nephropathy (CAN) is an iatrogenic complication of routine radiography with iodinated contrast media. Although the increase in serum creatinine is typically transient, CAN is associated with increased risks of mortality, major adverse cardiac events, and new onset or progression of chronic kidney disease. Few prophylactic therapies have proven effective in prevention of CAN. Sodium chloride and sodium bicarbonate infusions before and after contrast exposure are considered by many to have the most robust supportive evidence base. However, the impact of these therapies on the associated “downstream” adverse events has not been formally tested. In particular, the effects of sodium chloride or sodium bicarbonate on mortality have not been studied systematically. We sought to conduct a meta-analysis using the multiple randomized trials comparing intravenous sodium bicarbonate to sodium chloride for the prevention of CAN to determine if 1 therapy is more associated with a reduction in mortality or incidence of CAN. Because mortality is infrequent, we performed this meta-analysis using individual patient–level data rather than study level data. This would allow for more precision in determining time to death and the relation (if any) to CAN severity. We tested the null hypothesis that all-cause mortality hazards do not differ between patients randomized to periprocedural sodium bicarbonate compared to those randomized to periprocedural sodium chloride. We also tested whether any differences in the incidence of death were accompanied by differences in the incidence of CAN in the same trials.
Methods
Potentially eligible trials were identified through standardized electronic database searches for journal articles or meeting abstracts, in any language, indexed within MEDLINE, Web of Science, and/or BIOSIS from inception until June 12, 2014 ( Supplementary Table 1 ). Reference lists of included trials and other content-relevant journal articles were reviewed manually.
Included trials were determined by consensus according to predefined eligibility criteria: (1) population: patients undergoing coronary angiography; (2) intervention: intravenous isotonic sodium bicarbonate; (3) control: intravenous sodium chloride; (4) outcome: all-cause mortality—number and proportion of participants in both intervention and control arms; and (5) study design: double-blind, single-blind, or open-label randomized trials. Six trials that reported 0 deaths overall were excluded.
Individual patient data sets were requested from authors of included trials, standardized, and aggregated into a single data set. To verify accuracy, the number of deaths, CAN events, and participants in each arm were checked against original reports. As needed, we sought clarification from the relevant trialist; when data from the individual patient data set could not be reconciled with the original report for a trial, we used the former as the definitive data source. Variables consistently reported across trials were age, gender, history of diabetes mellitus, hypertension or congestive heart failure, left ventricular ejection fraction, contrast volume administered, Mehran risk score, baseline serum creatinine concentration and estimated filtration rate, all-cause mortality, cause of death, CAN, and time from randomization to death or censor. For included trials that did not provide individual patient data sets, time-to-event data were imputed on the basis of the trial or treatment arm–specific follow-up periods specified in the original reports ( Supplementary Table 2 ). We performed sensitivity analyses to assess the influence of including such trials. The trials’ original reports were used to abstract data on trial-level characteristics, including eligibility criteria (as mentioned previously), source (first author surname, year, PubMed identifier), design (blinding, parallel vs factorial, ineligible trial arms, eligibility trial arms involving confounded comparisons), setting (single center vs multicenter), procedure (coronary angiography, percutaneous coronary intervention [PCI]), contrast medium (agent, ionicity, osmolality), fluid administration protocol (concentration, rate, dose, duration), inclusion criteria (renal insufficiency, none), and criteria used to define CAN (end point assessment, biomarker changes).
The primary outcome, all-cause mortality hazard (instantaneous mortality risk), was defined as the time from randomization to death or censorship, whichever occurred first. Participants alive and still being followed up were censored at 1 year to analyze the primary outcome. The proportional hazard assumption was assessed to confirm that mortality hazards were generalizable from randomization to the end point. The secondary outcome, CAN risk, was defined according to the criteria provided within the original report of each trial. When >1 CAN outcome was reported, the primary outcome criteria were was preferentially used.
Survival functions were generated for sodium bicarbonate (intervention) and sodium chloride (control) arms based on time-to-event data in the pooled data set using Kaplan–Meier methodology. Primary outcome survival analysis was done using a 2-stage approach, as in earlier time-to-event individual patient data meta-analyses. In stage 1, primary outcome effect estimates were summarized as hazard ratios (HRs) and 95% confidence intervals (CIs) for each trial; these were calculated using nonparametric log-rank tests. In stage 2, trial-specific estimates were pooled in a meta-analysis using the Mantel–Haenszel fixed-effect model. We completed sensitivity analyses to assess the influence of the statistical method chosen to do the survival analysis, using several alternative approaches ( Supplementary Data ).
Dr. Brown, Dr. MacKenzie, and Pearlman analyzed the data. IRB approval was waived for this study because no patients were actively enrolled. We received a waiver from the Center for the Protection for Human Subjects (CPHS STUDY00029137).
Results
Results of systematic search and trial selection processes are summarized in the appendix ( Supplementary Figure 1 ). Ten trials (2,764 participants) met eligibility criteria ( Table 1 ). We excluded 1 trial arm each from 2 of these trials—one because participants did not receive any volume expansion infusion and another because participants received oral, not intravenous, sodium bicarbonate. Both of these trials initially had 3 treatment arms. Two of the 10 trials involved confounded comparisons, having compared preprocedural and postprocedural sodium bicarbonate to only postprocedural sodium chloride. In contrast to the aforementioned excluded arms, these were consistent with eligibility criteria and therefore retained in the analyses. The influence of this decision on main outcome was assessed by sensitivity analyses ( Supplementary Tables 3 and 4 ).
| BOSS (2014) | Brar (2008) | Gomes (2012) | Klima (2012) | Maioli (2008) | Maioli (2011) | Masuda (2008) | Recio-M. (2007) | Thayssen (2014) | Ueda (2011) | |
|---|---|---|---|---|---|---|---|---|---|---|
| Source | ||||||||||
| PubMed identifier | Abstract | 18768415 | 23184077 | 22267245 | 18702961 | 21972403 | 17719320 | 17394959 | 241714489 | 21349483 |
| Individual patient data | ● | ○ | ● | ● | ● | ● | ○ | ● | ● | ○ |
| Design | ||||||||||
| Open-label | ○ | ○ | ● | ● | ● | ● | ● | ○ | ○ | ○ |
| Single-blind | ○ | ● | ○ | ○ | ○ | ○ | ○ | ● | ● | ● |
| Double-blind | ● | ○ | ○ | ○ | ○ | ○ | ○ | ○ | ○ | ○ |
| Parallel, 1:1 | ● | ● | ● | ○ | ● | ○ | ● | ● | ○ | ● |
| Factorial, 1:1:1:1 ∗ | ○ | ○ | ○ | ○ | ○ | ○ | ○ | ○ | ● | ○ |
| We excluded 1 trial arm | ○ | ○ | ○ | ● | ○ | ● | ○ | ○ | ○ | ○ |
| Confounded comparison | ○ | ○ | ○ | ○ | ○ | ● | ○ | ● | ○ | ○ |
| Setting | ||||||||||
| Country | USA | USA | Brazil | Switzerland | Italy | Italy | Japan | Spain | Denmark | Japan |
| Accrual period | 2010–12 | 2006–07 | 2004–08 | 2005–09 | 2005–06 | 2004–08 | 2005–06 | 2004–05 | 2010–12 | 2008–10 |
| Single-centre | ○ | ● | ○ | ○ | ● | ● | ● | ● | ○ | ● |
| Index procedure † | ||||||||||
| Coronary angiography | ● | ● | ● | ● | ● | ○ | ● | ○ | ○ | ● |
| Other PCI | ● | ● | ● | ● | ● | ○ | ● | ● | ○ | ● |
| Primary PCI | ○ | ○ | ○ | ○ | ○ | ● | ○ | ○ | ● | ○ |
| Intention-to-treat (N) | ||||||||||
| Sodium bicarbonate | 146 | 175 | 150 | 37 | 250 | 150 | 29 | 56 | 358 | 30 |
| Sodium chloride | 145 | 178 | 151 | 35 | 252 | 150 | 30 | 55 | 357 | 30 |
| Contrast medium | ||||||||||
| Agent | Multiple ‡ | Ioxilan | Ioxaglate | Multiple § | Iodixanol | Iodixanol | Iopamidol | Iomeprol | Iodixanol | Multiple ¶ |
| Non-ionic | ● | ● | ○ | ● | ● | ● | ● | ● | ● | ● |
| Ionic | ○ | ○ | ● | ○ | ○ | ○ | ○ | ○ | ○ | ○ |
| High-osmolar | ○ | ○ | ○ | ○ | ○ | ○ | ○ | ○ | ○ | ○ |
| Low-osmolar | ● | ● | ● | ● | ○ | ○ | ● | ● | ○ | ● |
| Iso-osmolar | ● | ○ | ○ | ● | ● | ● | ○ | ○ | ● | ○ |
| Sodium bicarbonate protocol | ||||||||||
| Concentration (mEq/L) | 154 | 154 | 154 | 166 | 154 | 154 | 154 | 154 | 167 | 154 |
| Rate 1 (mL/kg/h) | 3 | 3 | 3 | 3 | 3 | 3 | 3 | 5 | 0 | 0.5 |
| Duration 1 (h) | 1 | 1 | 1 | 1 | 1 | 1 | 1 | 1 | 0 | NR |
| Rate 2 (mL/kg/h) | 1 | 1·5 | 1 | 1 | 1 | 1 | 1 | 1·5 | 1·4 | 1 |
| Duration 2 (h) | 6 | 4 | 6 | 6 | 6 | 12 | 6 | 12 | 5 | 6 |
| Sodium chloride protocol | ||||||||||
| Concentration (mEq/L) | 154 | 154 | 154 | 154 | 154 | 154 | 154 | 154 | NR | 154 |
| Rate 1 (mL/kg/h) | 3 | 3 | 3 | 1 | 1 | 0 | 3 | 0 | 0 | 0·5 |
| Duration 1 (h) | 1 | 1 | 1 | 12 | 12 | 0 | 1 | 0 | 0 | NR |
| Rate 2 (mL/kg/h) | 1 | 1·5 | 1 | 1 | 1 | 1 | 1 | 1 | NR | 0 ‖ |
| Duration 2 (h) | 6 | 6 | 6 | 12 | 12 | 12 | 6 | 12 | NR | 0 ‖ |
| Inclusion criteria | ||||||||||
| Renal insufficiency | ● | ● | ● | ● | ● | ○ | ● | ○ | ○ | ● |
| None | ○ | ○ | ○ | ○ | ○ | ● | ○ | ● | ● | ○ |
| Contrast-associated nephropathy | ||||||||||
| Endpoint assessment (h) | 24–72 | 24–96 | 48–48 | 0–48 | 0–120 | 0–72 | 0–48 | 0–72 | 48–72 | 0–48 |
| Δ sCr >25% increase | ● | ○ | ○ | ● | ○ | ● | ● | ○ | ● | ● |
| Δ sCr >0·5 mg/dL increase | ● | ○ | ● | ● | ● | ● | ● | ● | ○ | ● |
| Δ eGFR >25% decrease | ○ | ● | ○ | ○ | ○ | ○ | ○ | ○ | ○ | ○ |
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