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,2
* Department of International Health, Rollins School of Public Health, Emory University, Atlanta, GA 30322;
Program in Nutrition and Health Sciences, Graduate Division of Biological and Biomedical Sciences, Emory University, Atlanta GA 30322; and
** Department of Statistics, Purdue University, West Lafayette, IN 47907
2To whom correspondence should be addressed. E-mail: uramakr{at}sph.emory.edu.
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ABSTRACT |
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 TOP ABSTRACT MATERIALS AND METHODSRESULTSDISCUSSIONLITERATURE CITED |
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KEY WORDS: • vitamin A • iron • multimicronutrient • meta-analysis • growth
Child undernutrition, as indicated by linear growth failure and wasting, remains a major public health concern worldwide because of its significance for child morbidity and mortality and long-term consequences, such as reduced adult muscle mass and increased obstetric risk (1,2). Although substantial progress was made over the past few decades, stunting still affects about a third of preschool age children in developing nations, whereas wasting affects 
3, 7, and 10% of preschool children in Latin America, Africa, and Asia, respectively (3–5). Growth retardation usually begins in utero in many of these settings and continues during the first 2–3 y of life as a result of inadequate food intake and infections such as diarrhea (6). Research on the causes of growth failure over the last half century focused initially on protein and then energy intake; recently, more attention has been paid to micronutrients. Several randomized controlled trials (RCTs)3 were conducted in developing countries to examine the effects of many of these nutrients either alone or in combinations of 2 or more on a range of maternal and child health outcomes including early childhood growth and development (7–14).
The most conclusive evidence to date linking the intake of a specific micronutrient to child growth is for zinc. Brown et al. (15) concluded that zinc supplementation has a significant positive effect on both linear and ponderal growth of prepubertal children based on a meta-analysis of > 30 RCTs. A pooled analysis of RCTs also showed that zinc supplementation significantly reduced the incidence of diarrhea and pneumonia in preschool children (16), which are associated with poor growth. In contrast, the evidence is less clear for the role of vitamin A and iron on child growth. Although observational studies reported significant correlations between vitamin A status and stunting (17–19) and wasting (17,20), the results from RCTs are contradictory. In a recent review, Bhandari et al. (21) concluded that vitamin A supplementation had little effect on linear growth of young children, but the evidence was not evaluated using a meta-analysis and included nonrandomized studies. Although several observational studies showed a positive correlation between stunting and iron deficiency (22–24), the evidence that iron supplementation improves child growth is also contradictory, with some studies showing significant improvement, and others reporting null findings or even suggesting harmful effects. No systematic review of the findings of these studies has been conducted to date, making it very difficult to draw any overall conclusions from the current research on iron supplementation as it relates to child growth.
The main objective of this paper was therefore to compile the results of multiple studies and assess, through a formal meta-analysis, the overall effect of vitamin A and iron on the growth of children. Also, we assessed whether effects varied by baseline nutritional status, age, duration, dosage, and mode of intervention. In addition to these 2 nutrients, we also evaluated the effect of multimicronutrient interventions on child growth.
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MATERIALS AND METHODS |
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TOPABSTRACTMATERIALS AND METHODSRESULTSDISCUSSIONLITERATURE CITED |
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Potential studies for inclusion into the meta-analysis were identified through 2 searches on PubMed database 1966-present on September 11, 2003 for vitamin A, on August 24, 2003 for iron, and on October 21, 2003 for multimicronutrients. The first search for each intervention included the word "vitamin A" or "iron" or "micronutrient" in the title and the words "growth, infant or child or children" in any field and the second search included the word "supplement or supplemental" in the title and the words "growth or weight or length or height" in any field. Additional studies were identified through the bibliographies of review articles.
Exclusion criteria. Animal studies, review articles, cell or tissue studies and nonintervention studies were rejected. Additional exclusion criteria included studies in which 1) the treatment and control groups differed in more than simply their inclusion of vitamin A, or iron, or multimicronutrients; 2) participants were >18 y old; 3) there was no control group; 4) there were doubts about randomization; 5) there was a lack of sufficient data on growth to calculate an effect size; and 6) the duration of follow-up was <8 wk. The selection process resulted in a final set of published studies that were randomized, controlled intervention studies in children < 18 y old in which the intervention provided to treatment and control participants differed only in the inclusion of the micronutrient(s) of interest (vitamin A, or iron, or multimicronutrients).
Statistical analyses
The primary outcomes of interest were changes in height (cm) and weight (kg). Changes in serum retinol or hemoglobin (Hb) concentration were also examined for vitamin A and iron, respectively. Effect sizes were calculated for individual studies by dividing the difference between the mean change in treatment and control groups by the pooled SD. This value is known as Cohen’s effect size or Cohen’s d, and is useful in meta-analyses because it eliminates the problems of units of measurement and duration, which may vary among studies (25). The overall mean effect size and 95% CI across studies was then estimated assuming the random effects model that used the weighted mean effect size for each study in which the weight was the inverse of the intrastudy variance.
We tested for heterogeneity by using the 
2 test of homogeneity as described by Hedges (26) to test the hypothesis that the population effect sizes across studies were equivalent and by visual examination in which effect sizes calculated for each study were compared with the overall pattern of effect sizes. Outliers were defined as any study which differed markedly from the overall pattern. Overall weighted mean effect sizes were calculated with and without outliers. Several potential effect modifiers were also considered in the analyses. Studies were stratified according to age of participants, mode of administration, dosage, duration of intervention, and initial Hb (iron only) and Z-scores. Weighted mean effect sizes were calculated in each stratified subgroup that contained at least 2 studies. Regression analyses were also performed to determine whether study duration, baseline weight-for-age Z-score (WAZ), height-for-age Z-score (HAZ), weight-for-height Z-score (WHZ), and Hb or serum retinol levels predicted effect sizes.
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RESULTS |
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TOPABSTRACTMATERIALS AND METHODSRESULTSDISCUSSIONLITERATURE CITED |
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Study attributes. Of the 170 potential vitamin A studies identified, 156 were rejected and of the 39 intervention trials in children, 14 studies that yielded 17 datasets with growth outcomes were included in the meta-analysis and are described in Table 1. Because the publication by Kirkwood et al. (27) included the results of 2 distinct studies referred to by the authors as "Health" and "Survival," we included the results in 2 data sets which we called KirkwoodHealth and KirkwoodSurvival. Mwanri et al. (28) compared vitamin A with a placebo and compared vitamin A + iron with iron alone, resulting in 2 data sets. The study by Rahman et al. (29) compared vitamin A with placebo and vitamin A + zinc with zinc alone, also resulting in the 2 data sets. Vitamin A was provided as a high-dose supplement (liquid or capsule) in most of the studies (27,29–38), except for the following: 1) Mwanri et al. (28) in which a vitamin A–fortified corn gruel was compared with a nonfortified corn gruel; 2) Rahmathullah et al. (39) in which a weekly dose was used; and 3) Yang et al. (40) in which a supplement (vitamin A + calcium + zinc vs. calcium + zinc) was provided 5 d/wk either mixed with milk powder or in tablet form. The duration of the studies ranged from 3 to 24 mo and most had a follow-up period of 4–6 mo post-vitamin A dosage. The studies using high-dose vitamin A supplements provided 1–6 doses with 4–6 mo between doses. The studies by Mwanri et al. (28) and Bahl et al. (32) had a follow-up period of 3 mo, whereas those of Rahmathullah et al. (39) and Yang et al. (40) reported changes in anthropometric measures after 52 wk of weekly or daily supplementation. The initial mean age of the participants ranged from 9 mo to 10.5 y as follows: <2 y (3 studies), 2–5 y (10 studies), and >5 y (1 study).
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Iron
Study attributes. Of the 413 potential studies identified for the iron meta-analysis, 392 were rejected, and of the 59 intervention trials among children, 21 studies resulting in 28 data sets that had data on growth were included in the iron meta-analysis (Table 2). The study by Chwang et al. (41) separated the participants according to Hb status before randomization and reported the results for anemic and nonanemic participants each with a treatment and control group. In this case, each data set was included in the analysis separately. Similarly, in the study by Dewey et al. (42) there were 2 test sites, Honduras and Sweden, which yielded 2 data sets for the meta-analysis. Many studies included a deworming regimen. If groups differed by deworming treatment, only data comparing the iron and placebo with the same deworming status were compared (43–45). In the paper by Dossa et al. (43), there were 2 treatments and 2 controls (iron + albendazole vs. placebo + albendazole and iron + placebo vs. placebo + placebo); therefore 2 data sets were generated from this study. The study by Mwanri et al. (28) compared iron with a placebo and compared iron + vitamin A to a vitamin A only group also resulting in 2 data sets. Similarly, 3 studies (46–48) that used different combinations of iron and zinc also yielded 2 datasets each, i.e., comparisons of iron only to placebo and iron + zinc to zinc only.
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Height/Weight. Sufficient information for the calculation of effect size on change in absolute height was available in 19 studies, which yielded 26 data sets (Fig. 3) with 3444 participants. There were 21 studies for a total of 28 data sets and 3610 participants with sufficient data to calculate effect sizes for iron intervention on weight gain (Fig. 4). The effect sizes for the study by Majumdar et al. (56) were > 5 times greater than the next largest effect size [height –5.00 (95% CI: –5.80, –4.20), weight –8.14 (95% CI: –9.34, –6.94)] and therefore not included in Figures 3and 4; comparison of summary estimates, with and without them, indicated that their removal did not significantly affect the summary estimate but did reduce the 95% CI. The overall weighted mean effect size after excluding the outlier was 0.09 (95% CI: –0.07, 0.24) for height and 0.13 (95% CI: –0.05, 0.30) for weight.
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Hemoglobin. There were 16 studies with a total of 21 data sets and 2542 participants with sufficient information to calculate effect sizes of iron intervention on Hb concentrations. The effect sizes ranged from 0.00 to 14.53. The largest effect size, 14.53 (95% CI 12.17, 16.88), occurred among severely anemic children in the study by Chwang et al. (41) that was defined as an outlier. The overall weighted mean effect size without the above study was a significant value of 1.49 (95% CI: 0.46, 2.51).
Multiple micronutrients
Study attributes. The database searches on multimicronutrient intervention trials resulted in 225 potential studies. Six more studies were identified through other sources for a total of 231 potential studies, of which 226 were rejected. Almost a third (n = 74) were review papers/commentaries/editorials; a number of food supplementation trials were also excluded because they did not permit isolation of the effects of micronutrients. Of 18 potential trials conducted in children, only 5 RCTs in which the treatment and control differed only in the inclusion of multimicronutrients were included in the meta-analysis (Table 3).
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5 d/wk either as a syrup (62), fortified beverage (63,64), or a fortified food (65,66) and contained vitamin A, iron, Zn, B vitamins, and folic acid in all studies. Some interventions also contained iodine (62,64), vitamin C (63,66), vitamin E (63,64), calcium (63,65,66), potassium (63,66), copper (62,66), and other trace vitamins and minerals (62). Three studies had participants < 24 mo old (62,65,66), whereas 2 had participants > 5 y old (63,64). Height/Weight. Sufficient information for the calculation of effect size on change in absolute height was obtained in 4 studies totaling 1330 participants and for change in weight gain in 5 studies totaling 1604 participants (Table 4). The overall weighted mean effect size was 0.28 (95% CI: 0.16, 0.41) for height and 0.28 (95% CI: –0.07, 0.63) for weight, respectively. There was no evidence of heterogeneity in the case of linear growth.
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DISCUSSION |
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TOPABSTRACTMATERIALS AND METHODSRESULTSDISCUSSIONLITERATURE CITED |
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The lack of effect of vitamin A on growth is indeed contrary to the dramatic reductions seen in young child mortality and the severity of infections such as measles and diarrhea, which are known predictors of child growth. Many of the studies that examined child growth were done in settings similar to those that examined child morbidity and mortality. Although 3 studies reported significant effects of vitamin A supplementation on either linear growth or weight gain, the overall weighted mean effect size for vitamin A for both these outcomes was indeed small (<0.1) and not significant. Further, the overall conclusions did not differ with our without the inclusion of outliers. In terms of adverse effects, only 1 study (37) reported a significant negative effect size (–0.70) for height, but the same study had a significant positive effect on weight gain (1.15).
One limitation was the use of sample averages to classify studies for the stratified analysis especially when there was considerable variation within a study. For example, the initial age of participants in 2 studies ranged from 6 mo to 6 y (33,37), resulting in the study being classified in the age category of 2–4.9 y, even though there were 3 distinct categories, i.e., infants, toddlers, and older preschool-age children, who differ in their growth patterns and perhaps response to the intervention. The use of averages may also have limited the effectiveness of the regression analysis, which found no significant correlations between initial characteristics of participants and effect size. Nevertheless the findings of some studies that conducted post-hoc analyses for some of these characteristics are consistent with our conclusions. For example, West et al. (36) reported improved growth after vitamin A supplementation in boys aged 4–5 y but not in younger boys, whereas 5 other studies (29,30,32,35,39) did not find any differences in growth based on age. Similarly, no differences were seen by initial Z-scores of participants (29). An interesting finding, however, was the effect of season on the response to vitamin A. Supplementation of vitamin A during the summer months in India resulted in a significant weight gain in participants age 1–5 y, whereas supplementation during any other season during the year resulted in no changes in weight or height increment compared with controls (32). Significant seasonal variation in dietary intake of vitamin A is common, especially in settings in which the main sources of this vitamin are fruits and vegetables (67–70), and it is reasonable to expect vitamin A interventions to benefit children more during the time of year when they are at greatest risk of vitamin A deficiency (38). Although only the study by Bahl et al. (32) examined the role of season, it should be noted that many studies had at least 1 y of follow-up, which could have diluted any positive effect of vitamin A supplementation received during any one season of the year. Finally, another possible reason for the lack of effect is that most studies excluded children with clinical symptoms of vitamin A deficiency; although the effect on growth may be limited to this subgroup (29,30,32–34,36,71), reductions in mortality were demonstrated in children with mild-to-moderate vitamin A deficiency (8). The lack of data on baseline serum retinol limited our ability to examine the role of this factor.
The results of the meta-analysis of iron intervention trials were similar to those seen for vitamin A. In contrast to the lack of effect on growth, there was an overall effect on Hb, confirming the efficacy of iron interventions in reducing anemia. As with the vitamin A studies, the possible reason that the overall weighted mean effect sizes were not significant for the effect of iron on growth could be due to the heterogeneity of studies. Although we did not find any effects even after stratifying studies according to baseline characteristics and mode of intervention, there were considerable interstudy differences within the strata. The findings of the few studies that conducted post-hoc analysis, however, indicated that that there were no differences in the effect of iron supplementation after stratifying subjects based on initial HAZ (43,47,54), Hb status (43), or age (54). Only one study used a multiple regression approach to determine any effect modifiers and found that the positive effect of iron on growth was decreased with increasing age over the age range of 7–9 y (50).
There has been recent concern that widespread iron supplementation could actually have a detrimental effect on growth in iron-replete children and/or young infants (21). None of the studies that enrolled iron-replete participants with a mean initial Hb value > 110 g/L reported a significant negative effect of iron on growth, whereas 3 data sets had significantly positive effect sizes for height (52) and weight (50,52,58). Additionally, regression analysis found no significant negative correlation between effect size for either height or weight and initial Hb. A major contribution of this meta-analysis is that although there is no overall effect of iron interventions on growth, the findings suggest that there are no adverse effects in iron-replete children, thereby indicating that targeting all children for iron supplementation in the effort to eliminate iron deficiency and anemia is safe. However, we are limited by lack of adequate data on the safety of iron supplements for young infants (<6 mo of age).
In contrast to the lack of effect on growth of vitamin A or iron interventions when administered as single nutrients, our findings suggest that correcting multiple nutrient deficiencies simultaneously is efficacious. In many settings, growth faltering has been associated with overall poor dietary quality, which includes inadequate intakes of animal foods that are sources of highly bioavailable forms of several micronutrients, and high intakes of inhibitors such as phytates (72–75). We found that multimicronutrient interventions that provided the recommended daily allowance of iron, zinc, vitamin A, folic acid, and B vitamins had a positive effect on height and weight gain in children. Although there were a limited number of studies, the results were reported from Latin America, Africa, and Asia, had durations of 8–52 wk, and participants ranging in age from 6 mo to 11 y.
The increased effectiveness of multimicronutrient interventions compared with vitamin A or iron alone may be explained by the high prevalence of concurrent micronutrient deficiencies (68,74,76–79) and/or the positive synergistic effects between some of these nutrients at the level of both absorption and metabolism (for example, vitamin A and iron, vitamin A and zinc, iron and vitamin C) (74). Another plausible explanation is the role of zinc. A comparison of our results with the effect sizes reported for height (0.350, 95% CI: 0.189, 0.511) and weight (0.309, 95% CI: 0.178, 0.439) with zinc supplementation in the meta-analysis by Brown et al. (15) suggests that the effectiveness of the multimicronutrient interventions in improving growth may have been due to zinc, which was included in all interventions. Further, our results also suggest that recent concerns about the competitive interaction between iron and zinc (46,80) may not be a problem, at least for growth outcomes, when zinc is included in a multivitamin-mineral supplement.
In summary, interventions with a single nutrient such as vitamin A or iron, although providing benefits such as improved Hb status and reduced mortality, may not be the optimal for addressing growth failure, whereas interventions that provide multiple micronutrients and/or zinc may be the most effective in improving child growth. The feasibility of supplementation as the mode of delivery, however, remains a concern, especially in poor resource settings. More sustainable food-based approaches such as fortification, improving dietary quality, and education to improve micronutrient intakes of young children must be pursued.
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FOOTNOTES |
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3 Abbreviations used: BW, body weight; HAZ, height-for-age Z-score; Hb; hemoglobin; RCT, randomized controlled trials; WAZ, weight-for-age Z-score; WHZ, weight-for-height Z-score.
Manuscript received 16 March 2004. Initial review completed 16 May 2004. Revision accepted 3 August 2004.
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TOPABSTRACTMATERIALS AND METHODSRESULTSDISCUSSIONLITERATURE CITED |
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