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Background: Although various micronutrient regimens have been shown to prevent and treat common infectious diseases in children, the effects of daily multivitamin (MV) and/or zinc supplementation have not been widely evaluated in young African infants. Objective: The objective was to determine whether daily supplementation of HIV-unexposed Tanzanian infants with MVs or zinc reduces the risk of infectious morbidity compared with placebo. Methods: In a 2 × 2 factorial, double-blind, randomized controlled trial, 2400 infants who were 6 wk of age and born to HIV-negative mothers in a low-malaria setting were randomly assigned to receive daily oral supplementation of MVs (vitamin B complex and vitamins C and E), zinc, zinc + MVs, or placebo for 18 mo. Morbidity was assessed by study nurses at monthly visits and by physicians every 3 mo and/or when the child was acutely ill. Results: No significant differences were found in the percentage of nurse visits during which diarrhea, cough, or any other symptom were reported throughout the previous month when receiving either zinc or MVs. However, physician diagnoses of all types of diarrhea (RR = 0.88; 95% CI: 0.81, 0.96; P = 0.003), dysentery (RR = 0.84; 95% CI: 0.74, 0.95; P = 0.006), and acute upper respiratory infection (RR = 0.92; 95% CI: 0.88, 0.97; P = 0.0005) were significantly lower for infants supplemented with zinc than for those who did not receive zinc. Among the 2360 infants for whom vital status was obtained, there was a nonsignificant increase in all-cause mortality among infants who received zinc (HR = 1.80; 95% CI: 0.98, 3.31; P = 0.06) compared with those who did not receive zinc. MVs did not alter the rates of any recorded physician diagnoses or mortality. Neither zinc nor MVs reduced hospitalizations or unscheduled outpatient visits. Conclusions: Daily zinc supplementation of Tanzanian infants beginning at the age of 6 wk may lower the burden of diarrhea and acute upper respiratory infections, but provision of MVs using the regimen in this trial did not confer additional benefit.
The study was a randomized, double-blind, 2 × 2 factorial design trial that took place in periurban Dar es Salaam, Tanzania (clinicaltrials.gov {“type”:”clinical-trial”,”attrs”:{“text”:”NCT 00421668″,”term_id”:”NCT00421668″}}NCT 00421668). Mothers of potentially eligible infants were recruited into the study in 1 of 2 ways: 1) pregnant women ≤34 wk gestation presenting at 1 of 3 prenatal clinics in Dar es Salaam were informed about the study and consented prenatally or 2) women were recruited from the labor ward of Muhimbili National Hospital within 12 h of delivering a healthy singleton baby. In both cases, written informed consent was obtained and mothers were asked to present at a study clinic within 1–2 wk of delivery for HIV testing. Maternal HIV status was determined using 2 sequential ELISAs that used the Murex HIV antigen/antibody (Abbott Murex) followed by the Enzygnost anti-HIV-1/2 Plus (Dade Behring) or the Enzygnost HIV Integral II Antibody/Antigen (Siemens). Any discrepancy between the first and second ELISA was resolved by a Western blot assay. Consenting mothers who were confirmed to be HIV-negative were enrolled into the study and their infants were randomly assigned to 1 of 4 regimens between 5 and 7 wk of age. Infants of multiple births and infants with congenital anomalies or other conditions that would interfere with the study procedures were excluded. Birth characteristics were obtained immediately after delivery whenever possible. We used reference data from Oken et al. (11) to calculate the percentile of birth weight for each completed week of gestation and defined small-for-gestational age as ≤10th percentile. At the time of randomization, clinical examination was performed by a study physician, history of morbidity and infant feeding practices was conducted by a study nurse, infant blood was drawn for a complete blood count, and anthropometric measurements were performed. Institutional approval was granted by the Harvard T.H. Chan School of Public Health Human Subjects Committee, the Muhimbili University of Health and Allied Science Committee of Research and Publications, the Tanzanian National Institute of Medical Research, and the Tanzanian Food and Drugs Authority. Over the course of the study, a Data Safety and Monitoring Board met twice annually. Infants were randomly assigned in a factorial design to receive a daily oral dose of 1 of the following 4 regimens for 18 mo from the time of randomization: 1) zinc, 2) MVs, 3) zinc + MVs, or 4) placebo. The biostatistician in Boston prepared a randomization list from 1 to 2400 that used blocks of 20 and was stratified by study clinic. Capsules were packaged in a blister pack of 15 each and numbered boxes containing 6 blister packs were prepared containing the corresponding treatments. Each eligible infant was assigned the next numbered box of capsules at his/her respective site. The supplement used was an orange-flavored powder encapsulated in an opaque gelatinous capsule and was manufactured by Nutriset. All 4 regimens were field tested and the taste, smell, and appearance were found to be indistinguishable between groups. All study personnel and participants were blinded to treatment assignment for the duration of the study. From the time of randomization to 6 mo of age, infants received 1 capsule/d, and from 7 mo of age to the end of follow-up, 2 capsules were provided daily. For infants in the zinc group, the capsule contained 5 mg of zinc. For infants in the MV group, the capsule contained 60 mg of vitamin C, 8 mg of vitamin E, 0.5 mg of thiamine, 0.6 mg of riboflavin, 4 mg of niacin, 0.6 mg of vitamin B-6, 130 mg of folate, and 1 mg of vitamin B-12. Infants in the MV + zinc group received 1 capsule containing the micronutrients listed in both the MV and the zinc groups. For children 0–6 mo of age, these doses represented between 150% and 600% of the RDA or Adequate Intake, and for children 7–12 mo of age, the doses were equivalent to 200–400% of the RDA or Adequate Intake. Mothers were shown how to push the capsule through the back of the blister pack, open the capsule, decant the powder into a small plastic cup, mix the powder with 5 mL of sterile water, and administer the solution to the child orally. Our choice of supplement composition was based on several considerations including: 1) previous research that found deficiencies in vitamin B-12, folate, zinc, vitamin A, and vitamin E among breastfeeding women in South Africa (10) and, therefore, suggests that infant micronutrient status may be low in sub-Saharan Africa; 2) a previous clinical trial involving pregnant Tanzanian women that confirmed improved birth outcomes with supplementation of vitamin B complex, vitamin E, and vitamin C (12); and 3) findings from our previous trial of MV supplementation involving HIV-exposed children, which revealed lower rates of fever and vomiting among supplemented children (13). Dar es Salaam has been described as a malaria-endemic area, although studies have suggested that rates of malaria are declining (14). Nonetheless, owing to the potential that iron supplementation of nonanemic children may have adverse consequences in malaria-endemic regions (15), the MV supplement did not include iron. Mothers and children were followed from the time of randomization for 18 mo, until the child’s death, or until loss to follow-up. Mothers who were enrolled during pregnancy received standard prenatal care including anthropometric assessment, intermittent prophylaxis for malaria, tetanus toxoid immunization, deworming using mebendazole, anemia assessment, and iron-folic acid supplementation. During this follow-up period, mothers and children were asked to return to the study clinic every 4 wk for data collection and standard clinical care, including growth monitoring, immunizations, routine medical treatment for illnesses, and periodic vitamin A supplementation (100,000 IU at 9 mo and 200,000 IU at 15 mo). Children who were diagnosed with anemia were treated with iron supplementation. At ~12 mo of age, CD4 and CD8 T cell counts and percentages were measured from a subset of 428 children using FACSCalibur system (Becton Dickinson) in order to determine any potential immune effect of micronutrient supplementation. At each of the monthly follow-up visits study nurses assessed compliance by counting the number of unconsumed capsules, assessed infant feeding practices, and conducted a morbidity history with the aid of pictorial diaries that mothers were instructed to complete daily in order to document the occurrence of any of the following symptoms: diarrhea; common cold; cough; difficulty breathing; fever; refusal to eat, drink, or breastfeed; pus draining from ears; and vomiting. They also assessed symptoms that were present on the day of the visit, recorded vital signs (including measurement of the child’s temperature and respiratory rate and detection of chest indrawing), and inquired about the occurrence of any unscheduled clinic visits or hospitalizations in the past month. Diarrhea was defined as ≥3 loose or watery stools within a 24-h period. Rapid respiratory rate was defined as >50 breaths/min in infants 2–11 mo of age and >40 breaths/min among infants ≥12 mo of age. At baseline, every 3 mo, and/or when acute illnesses were noted by the study nurse a study physician conducted a physical examination, diagnosed illnesses, and provided necessary medical treatment. Physicians underwent regular training so that they used standardized diagnostic criteria and treatment guidelines consistent with the WHO and Tanzanian Ministry of Health and Social Welfare policies. For the purposes of the analysis of physician diagnosis, “any form of diarrhea” included persistent diarrhea, acute diarrhea, dysentery, and/or intestinal parasites. Acute upper respiratory infection was defined as pharyngitis or rhinitis (both without fast breathing or chest indrawing). Acute lower respiratory infection was defined as cough or difficulty breathing, rapid respiratory rate (based on the same definition described previously), and either a fever of >38.3°C or chest retractions. “Any form of respiratory infection” included acute upper respiratory infection, acute lower respiratory infection, pulmonary tuberculosis, or other causes of pneumonia. Children who missed their scheduled monthly follow-up appointment were visited at home and their vital status was confirmed through contact with immediate family members. In cases of child death, a verbal autopsy was performed to determine the cause of death. The cause of death forms were then coded by 2 independent pediatricians (KPM and CPD), and any differences were resolved by a third pediatrician. The primary outcomes of the study were the incidence of clinical symptoms of diarrhea and lower respiratory infection. Power was calculated for 1 factor in the factorial design, which represented either the zinc or MV arm. Based on findings from our previous trial, we estimated that the mean (SD) number of diarrheal and lower respiratory illnesses per child per year in the placebo group would be 3.4 (4.2) and 2.1 (1.0), respectively (16). After applying a 2-sided α-value of 0.025 to yield an overall type I error rate of 0.05 for each primary outcome and allowing for a 15% loss to follow-up and minimum power of 80%, we calculated that we would require 2400 subjects to detect a reduction of 18% in the mean number of diarrheal episodes per year. We then calculated that with this sample size we would have 90% power to detect an 8% reduction in the mean number of episodes of lower respiratory illness per year. Data were double entered using Microsoft Access software (Microsoft Corp.) at the central study site and then converted to SAS data sets and uploaded to a secured UNIX-based server for analysis. Intent-to-treat analyses were conducted according to a pre-established data analysis plan. Descriptive statistics were used to summarize baseline characteristics of the study population. Frequencies were reported for categorical variables and the mean ± SD for continuous variables. The χ2-test and ANOVA were used to detect any differences among treatment groups. We used generalized estimating equations with the log link, binomial variance, and exchangeable correlation matrix to compare the proportion of follow-up visits in which the illness symptom had occurred in the previous 4 wk between factors. For physician diagnoses that were made during routine visits every 3 mo or during unscheduled visits during acute illness episodes, the mean number of diagnoses over the follow-up period was compared between factors using Poisson regression. In both sets of analyses we introduced interaction terms to test for joint effects between the zinc and MV factors. We also tested for effect modification by each factor and sex and low birth weight. When the P-interaction term was <0.10, stratified analyses were performed. Although the study was not powered to detect differences in mortality, we used Cox proportional hazards modeling to explore differences in all-cause and cause-specific mortality by factor. Values in the text are means ± SDs unless otherwise indicated. All analyses were performed using SAS software (version 9.2; SAS Institute).
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