Folate and vitamin B12 share metabolic pathways and are important for DNA and protein synthesis and therefore cell growth and differentiation(Reference Shane and Stokstad1). Poor intake of either may accordingly be a contributing factor to poor growth worldwide. Many children in low- and middle-income countries have poor status of either or both of these nutrients(Reference McLean, de Benoist and Allen2); the extent to which it has any consequences for growth is not known.
Deficiencies of folate and vitamin B12 are often part of general malnutrition and may contribute to poor growth in children in many low- and middle-income countries(Reference Allen3–Reference Taneja, Bhandari and Strand5). The main source of vitamin B12 is animal-derived foods, which are expensive and for cultural and religious reasons often not eaten at all(Reference McLean, de Benoist and Allen2). Besides breast milk, the best sources of folates are dark green vegetables and legumes(Reference Allen3). It is likely that these foods are inaccessible or only eaten in small amounts by many in poor populations.
Previously, we conducted a factorial design, double-blind, randomised controlled trial (RCT) of two RDA of vitamin B12 and/or folic acid supplementation in 1000 North Indian children aged 6 to 30 months. In the first and main phase of the study, we demonstrated that daily vitamin B12 supplementation for 6 months improved linear growth in children who were stunted, wasted or underweight at baseline(Reference Strand, Taneja and Kumar6). Furthermore, vitamin B12 status at baseline was positively associated with linear growth in children who were not given vitamin B12 daily, but there were no such association in those who got vitamin B12. Other studies on the association between vitamin B12 status and growth and the effect of vitamin B12 supplementation on growth show conflicting results(Reference Strand, Ulak and Kvestad7,Reference Strand, Ulak and Hysing8) .
The liver can store both folates and vitamin B12, and the effect of supplementation may persist long due to improved stores(Reference Bailey and Caudill9,Reference O’Leary and Samman10) . The objective of this follow-up study was to estimate if there was any effect of the vitamin B12 and/or folic acid supplementation during early childhood after 6 years (when the children were 6 to 9 years old), and the extent to which markers of these B-vitamins in early childhood were associated with or modified the effect of supplementation. The subgrouping variables were pre-specified and were based on the results from the paper reporting the effects on growth in the first phase of the RCT(Reference Strand, Taneja and Kumar6).
Materials and Methods
Study settings, design and participants
We contacted children who previously participated in a factorial designed, randomised, double-blind, placebo-controlled trial on the effect of vitamin B12 and/or folic acid supplementation on childhood infections and growth in New Delhi, India, from January 2010 to September 2011(Reference Taneja, Strand and Kumar11,Reference Winje, Kvestad and Krishnamachari12) . Details on the baseline study procedures have previously been published(Reference Taneja, Strand and Kumar11). Briefly, the children aged 6 to 30 months living in low- to middle-resource settings in New Delhi were randomised in blocks of 16 (in a 1:1:1:1 ratio) to receive either placebo, two RDA of vitamin B12, two RDA of folic acid or two RDA of both vitamins daily for 6 months. The randomisation list was generated in Stata by an individual not involved in the study. The allocation to the treatment groups was concealed, and the study participants, the study team and the involved scientists were blinded with regard to the group identity throughout the study period. The doses (approximately two RDA) for children aged above 12 months were 150 mcg of folic acid and 1·8 mcg of vitamin B12 (as cyanocobalamin). For children aged 12 months and below, the doses were 75 mcg and 0·9 mcg, respectively. The vitamins were delivered in a lipid-based nutritional supplement.
From September 2016, we approached the 1000 children from the baseline study and were able to get in contact with 798 of whom 791 consented to participate in this follow-up study (Fig. 1). All families were initially contacted by phone and invited to participate. A physical visit was made to the family’s address if no contact could be made. We requested the families who had moved out of the study area to come to the study clinic for a 1-d follow-up assessment. On the day of assessment, consent was taken from the children’s caregiver for participation in the study. Thereafter, we gathered information on socio-demography and the family situation through structured questionnaires.
Fig. 1. The participant flow in the study.
The trial was first registered at www.clinicaltrials.gov as NCT00717730 in July 2008 and at www.ctri.nic.in as CTRI/2010/091/001090 in August 2010. The follow-up study was then registered as CTRI/2016/11/007494 in November 2016.
This study was conducted according to the guidelines laid down in the Declaration of Helsinki, and all procedures involving human subjects were approved by the Society for Applied Studies Ethics Review Committee [SAS/ERC/VitB12/2016] and Norwegian Regional Committee for Medical and Health Research Ethics (REK VEST) [2014/1359]. Written informed consent was obtained from the caregivers of the subjects.
Assessments
Height was measured by Seca 213, to the nearest 0·1 cm twice in each child by a trained team of fieldworkers. Inter- and intra-observer standardisation exercises for anthropometric assessments were conducted before study initiation in the outcome ascertainment team, and these were repeated every 3 months throughout the study duration.
We obtained 3 ml of blood samples from all children at enrolment in a tube containing EDTA (Becton Dickinson). The plasma was centrifuged at approximately 1000 × g at room temperature for 10 min, separated, and transferred into storage vials and stored at −20°C until analysed. Microbiological assays estimated plasma concentration of vitamin B12 and folate(Reference Kelleher, Walshe and Scott13,Reference O’Broin and Kelleher14) . Total plasma homocysteine (tHcy), a sensitive and functional marker for both vitamin B12 and folate deficiency, was analysed using commercial kits (Abbott Laboratories, Abbott Park)(Reference Shipchandler and Moore15).
Statistical analyses
The original study included 1000 children. For this follow-up, we assumed that we would be able to re-enrol 800 children. With this sample size and assuming a sd of 1, we had 80 % power to identify an effect size of 0·2 height-for-age z-scores (HAZ) with a statistical significance of 0·05. We used the command ‘power twomeans’ in Stata for the sample size calculation. This calculation assumed no interaction between the two dimensions of the factorial RCT (folic acid and vitamin B12 supplementation).
Proportions, means (sd) or medians (interquartile range) were calculated for categorical and continuous variables by the four supplementation groups: vitamin B12, folic acid, vitamin B12 and folic acid and placebo for the baseline and follow-up characteristics of the children and their families. Children’s HAZ at follow-up was calculated using WHO growth standards(Reference de Onis16). The wealth of the families was determined by a wealth index created using principal component analysis based on assets owned by the household(17). Using the scores from the principal component analysis, the population was divided into five equal wealth quintiles: poorest, very poor, poor, less poor and least poor.
Based on the RCT design, we estimated the effect of vitamin B12, folic acid and vitamin B12 and folic acid supplementation on linear growth (HAZ and stunting) at follow-up in crude and multivariable regression models. In the multivariable models, we adjusted for the wealth quintile, since this quintile was not evenly distributed between the intervention arms. We also merged the children receiving vitamin B12 into a ‘vitamin B12’ group (n 385) and compared this group with the children who did not receive vitamin B12 (‘placebo B12’ group, n 403). Similarly, we compared the children receiving folic acid (‘folic acid’ group, n 388) with the children who did not receive folic acid supplementation (‘placebo folic acid’ group, n 406). We measured the effects of supplementing vitamin B12 or folic acid in various pre-defined subgroups. We used the same subgroups (age at enrolment, sex, breast-feeding status, baseline anthropometric status, and cobalamin, folate and tHcy concentration at enrolment) as in the manuscript that reported the effects of vitamin B12, folic acid and vitamin B12 and folic acid on growth immediately after the end of supplementation(Reference Strand, Taneja and Kumar6). We undertook these subgroup analyses unadjusted and adjusted for baseline age, wealth quintile, breast-feeding status, sex, as well as HAZ and weight-for-age z-scores. In these models, we also included interaction terms to measure whether the effects of vitamin B12 or folic acid were significantly different between the subgroups. We also measured the interaction between vitamin B12 and folic acid supplementation. Only one-way interactions were included in the statistical models.
In an observational (cohort) design, we examined the association between markers of vitamin B12 and folate (plasma cobalamin, folate and tHcy concentrations) at enrolment and HAZ and the proportion stunted at follow-up in multivariable regression models. To identify and adjust for confounders in these models, we used a method of purposeful selection of covariates(Reference Bursac, Gauss and Williams18,Reference Hosmer, Lemeshow and Sturdivant19) . The candidate variables were age and sex of the child, breast-feeding status at baseline, maternal and paternal years of schooling and wealth quintile. We also examined the interactions between vitamin B12 supplementation (vitamin B12 group) and cobalamin and tHcy concentrations, as well as the folic acid supplementation (folic acid group) and the folate and tHcy concentrations on linear growth (HAZ and stunting). The plasma concentrations of cobalamin, folate and tHcy were log (base 2)-transformed before used in the statistical analyses. We used generalised linear model with the Poisson distribution family and log link for dichotomous outcomes and the Gaussian distribution family and identity link for the continuous outcomes(Reference Zou20). These statistical analyses were performed in Stata, version 16 (Stata corporation).
We also used generalised additive models in the statistical software R version 3.3.3 (The R Foundation for Statistical Computing) to explore non-linear associations between cobalamin, folate and tHcy concentrations at baseline and HAZ at follow-up adjusting for the same potential confounders as in the generalised linear models(Reference Wood21). In these models, we present the association stratified by those who were and those who were not supplemented by vitamin B12. This stratified analysis was undertaken because of the interaction between vitamin B12 supplementation and vitamin B12 status.
Results
Figure 1 shows the flow of the participants through the study. Of the 1000 children in the main study, 791 children consented to participate and were included in the follow-up study (Fig. 1). Demographic characteristics of the 791 participants at follow-up by the four intervention groups are shown in Table 1. Mean (sd) age at follow-up was 7·4 (0·7) years, ranging from 6 to 9 years.
Table 1. Demographic information of 791 North Indian children at baseline and follow-up
(Number and percentages)
* Data available in 790 children at follow-up.
† Data available in 791 at follow-up.
‡ Data available in 785 children at follow-up.
The mean (sd) and median (interquartile range) HAZ scores and the proportion of children stunted by intervention group are shown in Tables 2 and 3. Children who had been randomised to receive vitamin B12 were somewhat taller than the other children (mean difference of 0·12 HAZ (CI −0·01, 0·25), corresponding to a difference of 0·7 cm (CI −0·15, 1·57)); however, these differences were not statistically significant.
Table 2. HAZ scores in children of 6–9 years old according to intervention groups
(Coefficient values and 95 % confidence intervals; mean values and standard deviations)
HAZ, height-for-age z-scores.
* Generalised linear model with the Gaussian family and identity link.
† Adjusted for wealth quintiles.
Table 3. Proportion of children stunted at 6–9 years old according to intervention groups
(Risk ratio and 95 % confidence intervals)
* Generalised linear model with the Poisson family and log link.
† Adjusted for wealth quintiles.
In the subgroup analyses, baseline cobalamin concentration modified the effect of vitamin B12 supplementation on growth (P for interaction < 0·001). These analyses showed that there was an effect of vitamin B12 supplementation in those who had a baseline cobalamin concentration < 200 pmol/l. We also observed statistically significant effects in other subgroups (Fig. 2); however, the interaction term between these subgrouping variables was small and did not reach statistical significance.
Fig. 2. The effect of vitamin B12 or folic acid supplementation in early life on linear growth 6 years later by subgroups.
The predictors for linear growth are shown in Table 4. The model shows that vitamin B12 supplementation modified the association between the cobalamin concentration at baseline and the HAZ score (P for interaction < 0·001). Log cobalamin concentration at baseline predicted the HAZ score (regression coefficient 0·26; 95 % CI 0·15, 0·38) in the children who did not receive vitamin B12 supplementation, but not in the children who received vitamin B12 supplementation (–0·01; 95 % CI –0·13, 0·10). There was no such interaction between folic acid supplementation and folate concentration at baseline. There were no interactions between the ‘vitamin B12’ supplementation group and ‘folic acid’ supplementation group and tHcy concentration on HAZ (P for interaction 0·230 and 0·953, respectively). The tHcy concentration at baseline was negatively associated with the HAZ scores (–0·15; 95 % CI –0·27, –0·01). The most saturated model with HAZ as outcome explained 20 % of the variability of the HAZ score at follow-up.
Table 4. Predictors for linear growth in children of 6–9 years old
(Adjusted coefficient and 95 % confidence intervals, n 791)
HAZ, height-for-age z-scores.
* log (base 2)-transformed value.
† P < 0.05.
The association between cobalamin concentration at baseline and stunting was modified by vitamin B12 supplementation (P for interaction = 0·004) (Table 4). In this analysis, the log-transformed cobalamin concentration at baseline was associated with stunting in the children who did not receive vitamin B12 supplementation (Relative Risk (RR) 0·70; 95 % CI 0·53, 0·93), but not in the children who did (RR 1·25; 95 % CI 0·90, 1·74). We did not see such interaction between the ‘folic acid’ supplementation group and folate status at baseline. Breast-feeding and wealth quintiles were found to be predictors for both HAZ and stunting at follow-up (Table 4).
The association between plasma cobalamin, folate and tHcy concentrations at baseline and HAZ scores at follow-up is depicted in Fig. 3. The figures show that HAZ scores increase with increasing log cobalamin concentration in the children who did not receive vitamin B12 supplementation, but not in the children who received vitamin B12. There was no association between log folate status at baseline and HAZ score at follow-up, but the HAZ scores decreased with increasing log tHcy levels at baseline. The generalised additive model plots did not reveal any non-linear associations.
Fig. 3. The association between plasma cobalamin, folate and tHcy during early childhood and linear growth 6 years later. The y-axis denotes the HAZ scores centred around the mean, and the x-axis denotes baseline plasma cobalamin, tHcy and folate concentrations. The graphs were constructed using generalised additive models in R, and the solid line depicts the association of plasma cobalamin, folate and tHcy during early childhood and HAZ 6 years later. The shaded area spans the 95 % CI of these associations. HAZ, height-for-age z-scores.
Discussion
In the original trial, reporting the short-term effects, we found that 6 months of vitamin B12 supplementation resulted in a modest but statistically significant effect on weight-for-age z-scores(Reference Strand, Taneja and Kumar6). We also found an effect of vitamin B12 supplementation on linear growth in subgroups of children who were stunted, underweight or wasted at baseline. In the current follow-up study, we found that vitamin B12 supplementation resulted in a small (statistically insignificant) overall effect on linear growth after 6 years. There was also a slight reduction in the risk of stunting in the vitamin B12 group. The 95% CI, however, of this measured effect were consistent with a 49 % reduction to a 5 % increase in the risk of stunting compared with placebo.
The subgroup analyses found that vitamin B12 status at baseline modified the effect of vitamin B12 supplementation on linear growth, demonstrating a substantial effect in children with baseline subclinical vitamin B12 deficiency but not in those with adequate status. Similarly, when using data in a prospective cohort design, the association between baseline cobalamin concentration was modified by vitamin B12 supplementation. The association between baseline vitamin B12 status (pre-enrolment plasma cobalamin concentration) and linear growth was observed only in children who were not supplemented with vitamin B12. In the children who did not receive vitamin B12 supplement (placebo B12 group), each unit increment in log (base2) plasma cobalamin concentration (i.e. a doubling of the concentration) was associated with a 0·26 increase in HAZ score and a 30 % reduction in the risk of being stunted after 6 years. This interaction between vitamin B12 status and supplementation suggests a causal effect of vitamin B12 supplementation on growth. We believe that we can suggest causality from this observational finding, because the effect modifier (i.e. supplementation) is a variable that has been randomised and at the same time affects vitamin B12 status.
Vitamin B12 is involved in two biochemical reactions in humans(Reference Molloy22). One of these, the methionine cycle is also dependent on an adequate supply of folate for the remethylation of homocysteine to methionine. Disruption of this cycle increases homocysteine and will affect gene regulation, DNA synthesis and, subsequently, growth(Reference Molloy22). Incomplete methylation can also limit linear growth through senescence of the resting zone chondrocytes of the growth plates(Reference Nilsson, Mitchum and Schrier23).
The results from our analyses indicate that folate is not a growth-limiting nutrient in this population. This is also supported by the fact that only plasma cobalamin, and not folate concentration, was negatively associated with plasma tHcy concentration at baseline. In the other metabolic pathway where vitamin B12 takes part, the B-vitamin acts as an enzymic cofactor for methylmalonyl CoA mutase, which is involved in the catabolism of certain fats and amino acids that could be important for growth. This reaction is not linked to folate metabolism, and an effect of vitamin B12 through this pathway could also explain the lack of interaction between folic acid and vitamin B12 supplementation.
Our findings from the observational design are in line with a recently published study from Nepal where vitamin B12 status estimated by cobalamin, methylmalonic acid and tHcy concentration in infancy was associated with linear growth 5 years later(Reference Strand, Ulak and Kvestad7). In this cohort study, infant status and maternal vitamin B12 intake and status during lactation were positively associated with growth when the children were around 5 years of age. However, the results from another study, also from Nepal, are at odds with the findings from the RCT design-results of this study(Reference Strand, Ulak and Hysing8). In this latter study, 2 µg of cyanocobalamin daily from infancy and for 1 year resulted in a metabolic response reflecting improved vitamin B12 status but had no effect on growth. The discrepancy between these RCTs can be explained by the many differences between the study populations. The daily dose and age of enrolment were reasonably similar between the studies. However, the study in Nepal restricted enrolments to children who were mildly stunted (< –1 HAZ upon enrolment). By focusing on a marginalised group of children from a relatively poor community, there could have been other health or nutritional factors that limited the response of supplementation on growth. It should be noted that both RCTs demonstrated a beneficial effect of vitamin B12 supplementation on the metabolic profile, which, in turn, may have implications for health. Similar to our findings, other studies from low- and middle-income countries have showed that mothers’ years of education, socio-economic and breast-feeding status were independent predictors of stunting(Reference Amugsi, Dimbuene and Kimani-Murage24,Reference Velusamy, Premkumar and Kang25) . This is an important reminder of the many factors that are important for healthy growth, all of which could modify the effect of supplementation of single nutrients.
There are some limitations to the study. First, the dosage might have been too small and the duration of supplementation too short. Second, the absorption of the vitamins may have been interfered by bacterial overgrowth(Reference O’Leary and Samman10). Third, impaired growth can also be caused by deficiencies of other growth-limiting nutrients, repeated infections and poor dietary quality(Reference Prendergast and Humphrey26). Fourth, no dietary assessments were done participants and parents. Some of these limitations may have underestimated the potential effect by optimising vitamin B12 or folate status. In other words, we might have overlooked important effects because the status was not improved sufficiently or because of other factors that affected growth. However, we do not believe that these factors could have attenuated our effect estimates substantially because we found a large effect of the supplementation of vitamin B12 in those with inadequate status. Lastly, the 20 % attrition rate is a limitation and could be a source of bias. However, there were no major differences between the children who were included in the follow-up and those who were not on the relevant demographic characteristics.
Successful inclusion of 80 % of the children from the baseline trial after approximately 6 years, high quality and robust assessment of growth, and the availability of different plasma nutrient biomarker concentrations are the strengths of the study. Another strength of the study is the randomised, double-blind design of the original trial enabling causal inferences from the observed results.
The results of this study could have important public health implications. Individuals who do not consume animal foods may benefit from vitamin B12-fortified foods or oral vitamin B12 supplements.
Conclusion
Our findings show that vitamin B12 deficiency limits growth and that vitamin B12 supplementation improves growth and reduces the risk of stunting in North Indian children with subclinical vitamin B12 deficiency. Being a growth-limiting nutrient in young children in New Delhi, vitamin B12 supplementation should be considered.
Acknowledgements
The authors thank Shruti Bisht for coordinating the study. The authors also thank Kiran Bhatia for support on the data management. Finally, the authors thank Mari Manger, Chittaranjan Yajnik and Helga Refsum for their contribution to the original trial. The Society for Applied Studies acknowledges the core support provided by the Department of Maternal, Newborn, Child and Adolescent Health, WHO, Geneva (WHO Collaborating Centre IND-158) and the Centre for Intervention Science in Maternal and Child Health (CISMAC; project number 223269), funded by the Research Council of Norway through its Centers of Excellence scheme and the University of Bergen, Norway.
This work was supported by Thrasher Research Fund (grant no. 02827) and the Research Council of Norway provided financial support for the original (grant no. 172 226) and the follow-up study (grant no. 234 495).
S. T., R. C., I. K., N. B. and T. A. S. designed the research; S. T., R. C. and T. A. S. conducted the research; S. T., R. C. and T. A. S. analysed the data or performed statistical analysis; and S. T., R. C., I. K., N. B. and T. A. S. prepared the manuscript. All the authors were responsible for the final content of this manuscript.
There are no conflicts of interest.
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