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Ready-to-use-therapeutic foods are an important component of the effective outpatient treatment of severe wasting, however, their effectiveness in the population-based prevention of moderate and severe wasting has not been evaluated.
To evaluate the effect of a 3-month distribution of ready-to-use-therapeutic food on the nutritional status, mortality and morbidity of children 6 to 60 mo of age.
A cluster randomized trial of 12 villages in Maradi, Niger. Six villages were randomized to intervention and 6 to no intervention. Villages were visited monthly from August 2006 to March 2007. All children in the study villages between 6 and 60 mo of age were eligible for recruitment.
Children with weight-for-height ≥ 80% of the NCHS reference median in the 6 intervention villages received a monthly distribution of one packet per day of ready-to-use-therapeutic food (500kcal / day) from August to October 2006. Children in the six non-intervention villages did not receive preventive supplementation with RUTF and comprised the control group. Active surveillance for conditions requiring medical or nutritional treatment was conducted in all twelve study villages throughout follow-up from August 2006 to March 2007.
Changes in weight-for-height Z score according to the World Health Organization Child Growth Standards and incidence of wasting (weight-for-height Z < −2) over 8 mo of follow-up.
The number of children with height and weight measurements in August, October, December and February was 3,166, 3,110, 2,936 and 3,026, respectively. The adjusted effect of the intervention on weight-for-height Z score from baseline to the end of follow-up was 0.22 Z (95% CI: 0.13, 0.30). The absolute rate of wasting and severe wasting, respectively, was 0.17 events per child-year (140 wasting events / 841 child-years) and 0.03 events per child-year (29 wasting events / 943 child-years) in the intervention villages, as compared to 0.26 events per child-year (233 severe wasting events / 895 child-years) and 0.07 events per child year ( 71 severe wasting events / 1,029 child-years) in the non-intervention villages. The intervention thus resulted in a 36% (95% CI: 17% – 50%) reduction in the incidence of wasting and a 58% (95% CI: 43% – 68%) reduction in the incidence of severe wasting. The mortality rate was 0.007 deaths per child-year ( 7 deaths / 986 child-years) in the intervention villages and 0.016 deaths per child-year (16 deaths / 1,099 child-years) in the non-intervention villages, resulting in a non-significant 49% (95% CI: 0.25, 1.05) reduction in mortality.
Short-term supplementation of non-malnourished children with ready-to-use-therapeutic food reduced the decline in weight-for-height Z score and the incidence of wasting and severe wasting over 8 mo.
Wasting (weight-for-height Z score (WHZ) < −2) affects approximately 10% of the world’s children under 5 yrs 1 and is an important contributor to the population attributable risk of child mortality and overall burden of disease.2 New outpatient and community-based models for the treatment of wasting have been shown to be effective in the rehabilitation of severely wasted children.3, 4 These models are made possible largely with the use of ready-to-use-therapeutic foods (RUTF). RUTF are energy-dense, micronutrient-enriched pastes that have a nutritional profile similar to the traditional F-100 milk-based diet used in inpatient therapeutic feeding programs and are often made of peanuts, oil, sugar, and milk powder.5, 6
RUTF has been shown to be effective in the treatment of severe and moderate wasting 6–8 and was associated with higher recovery and rates of weight gain among children at nutritional risk presenting to health centers in Malawi.9 The effectiveness of RUTF in the population-based prevention of moderate and severe wasting in children has not been previously evaluated.
Using data collected in a cluster randomized trial, this study aimed to assess the effect of a 3-month distribution of RUTF to non-malnourished children in a region with traditionally high levels of child malnutrition. A cluster randomized trial, with the village as the unit of randomization, was used given the study’s aim to evaluate the effectiveness of a population-based, preventive distribution of RUTF delivered at the village- rather than the individual- level. The primary hypotheses were that village-level supplementation with RUTF in the months preceding the annual harvest would prevent declines in individual weight-for-height and reduce the incidence of wasting in children 6 to 60 mo of age over a period of 8 mo. Because RUTF may have additional health effects, the intervention effect on individual height-for-age, stunting, mortality and morbidity from malaria, diarrhea, and respiratory infection were also examined.
Niger is a landlocked country of the Sahel with a population of approximately fourteen million people.10 Household food production is linked to rain-fed agriculture, where staple crops such as millet and sorghum are harvested once per year from September to October. Each year, the decrease in food quantity and quality experienced in the months preceding the harvest (August to October) is associated with an increase in wasting among children under 5 yrs. Maradi, located in the south-central part of the country bordering Nigeria, has some of the highest rates of malnutrition in the country.11 The prevalence of wasting (WHZ <-2 of the National Center for Health Statistics (NCHS) reference median) was estimated to be 11.6% in Maradi between January and May 2006 11.
Since 2001, Médecins sans Frontières (MSF) has provided treatment for severe wasting in Maradi at no cost in collaboration with the Ministry of Health of Niger. The therapeutic feeding program uses an outpatient approach to the treatment of malnutrition, through which children without serious complications are offered home-based treatment with the provision of RUTF. In 2006, treatment was extended to all moderately wasted children (weight-for-height < 80% of the NCHS reference median) < 5 yrs with the aim of preventing the presentation of severe wasting.
There are a total of 212 and 323 villages in the Madarounfa and Guidan Roumdji districts in the Maradi region, respectively. Villages eligible for inclusion in the study were those that had between 100 and 200 children 6 to 60 mo of age according to the most recent Niger census,12 experienced a prevalence of wasting ≥ 15% in 2005 according to admission records to local therapeutic feeding programs, and were not crossed by main (i.e. paved) roads. Fourteen villages in Madarounfa and 13 villages in Guidan Roumdji initially met the inclusion criteria, but 15 (8 in Madarounfa and 7 in Guidan Roumdji) of these were removed after a field visit for verification. Therefore, a total of 12 villages (6 in each district) were identified that met the above criteria (Figure 1). The leaders of all 12 eligible villages were informed of the study objectives and protocol and agreed to participate.
The unit of randomization was the village, and intervention assignment was stratified by district and made through the random selection of village names from a hat. Selection was made by a member of the field team not involved in the identification of eligible villages. The first 3 villages drawn from the 6 eligible in a district were assigned to the intervention group. The remaining 3 villages in each district were assigned to the non-intervention group. Thus, a total of 6 villages were assigned to the intervention group and 6 villages to the non-intervention group. Assignment was not blinded due to the nature of the intervention.
Follow-up was conducted in the study villages on a monthly basis from August 2006 to March 2007. Children between 6 and 60 mo of age during the follow-up period were eligible for inclusion. Children < 6 mo at the start the study but reaching 6 mo of age during the follow-up period were recruited when eligible, while children reaching 60 mo of age were removed from follow-up when no longer eligible.
Children with weight-for-height ≥ 80% of the NCHS reference median in the 6 intervention villages received a monthly distribution of 1 sachet per day of RUTF (92g, 500kcal / day, Plumpy’nut® from Nutriset, Malaunay, France) from August to October 2006. Distributions of the preventive supplement were made by field teams of trained nutritional assistants and research nurses and took place at the same time as the study’s active surveillance activities (described below). Children in the 6 non-intervention villages did not receive preventive supplementation with RUTF and comprised the control group.
Active surveillance for conditions requiring medical or nutritional treatment was conducted in all children 6 to 60 mo of age in the 12 study villages throughout follow-up from August 2006 to March 2007. Irrespective of intervention assignment, any child found with weight-for-height < 80% of the NCHS reference median at a follow-up visit was referred to the nutritional programs or health centers in the area for treatment provided at no cost. Outpatient treatment in the local nutritional programs consisted of 2 sachets per day of RUTF (1,000 kcal / day) and weekly follow-up until discharge. Inpatient care was made available to children who presented with medical complications. Other medical conditions identified during follow-up were referred to the neighboring governmental health facility. Treatment for malaria and non-complicated diarrheal diseases were provided during the follow-up visit, if indicated.
Surveillance activities, including anthropometric measurements and physical examinations, were conducted by field teams of trained nutritional assistants and research nurses in a dedicated central location in each village identified by the head of the village and field teams. Care givers were asked to accompany their children to these sites for follow-up on a monthly basis. At the first visit, we administered a standardized questionnaire to obtain information on household, maternal and child socio-demographic characteristics, child health history and feeding practices. We estimated child age at recruitment using a special event calendar if exact date of birth was unknown. An abridged questionnaire was used at each follow-up visit to obtain information on the major health events and feeding practices in the previous month.
At all visits, we measured child length/height and weight. Trained nutrition assistants carried out anthropometric measurements with the use of standardized methods and calibrated instruments. Child height (recumbent length if < 85 cm) was measured to the nearest 0.1 cm using a wooden measurement board. Weight was measured to the nearest 0.1 kg using a hanging Salter scale.
The presence of malaria, respiratory infection, and diarrhea was determined by trained research nurses during the physical examinations and interviews with the mother. A malaria HRP2 rapid diagnostic test (Paracheck-Pf®, Orchid Biomedical Laboratories, Goa, India) was used in children with fever to diagnose plasmodium falciparum infection . Respiratory infection was defined as cough or difficulty breathing within the last 3 days, as reported by the mother. Diarrhea was defined as >3 loose stools within the last day, as reported by the mother. If a child did not present for a study visit in the village, the head of village provided the cause of absence. If a child had died, the cause of death was provided by a family member or the head of village.
When the proportion of children absent per village exceeded 5%, we scheduled additional study visits to facilitate complete measurements on all children. All the information was collected on standardized forms and double entered into a computer database (EpiData, EpiData Association, Odense, Denmark).
Our primary study outcome measures were individual WHZ score according to the World Health Organization Child Growth Standards 13 and wasting (WHZ < −2). Our secondary measures included severe wasting (WHZ < −3), height-for-age Z (HAZ) score according to the World Health Organization Child Growth Standards, stunting (HAZ < −2), severe stunting (HAZ < −3), mortality, and prevalence of malaria, diarrhea, and respiratory infection. To detect a difference of 50% in the incidence of wasting between groups with 90% power (at the two-sided 5% level), accounting for a design effect of 2 due the cluster design and 15% loss to follow-up, we calculated that we would need to include 1,000 children in each group. Analyses were by intention-to-treat. All children from villages initially assigned to the intervention group (or non-intervention group) were analyzed as from the intervention group (or non-intervention group).
To verify the randomization assumption, we compared the prevalence of baseline characteristics between intervention groups using generalized estimating equations to adjust standard errors for clustering at the village-level. We fitted mean WHZ and HAZ curves using mixed effects models with restricted cubic splines.14 Knots were placed at 1, 2, 3, 4, and 5 months from the start of study. The outcome in each model was WHZ or HAZ and covariates included the intervention group, linear and spline terms for time (in months), and interaction terms between intervention group and time. We adjusted for child’s age at recruitment, sex, baseline HAZ, district, and interaction terms between these variables and time. Baseline WHZ and its interaction with time also were adjusted for when WHZ was the dependent variable. Baseline WHZ and HAZ were entered into the model as continuous terms. The mixed models used hierarchal random effects for the village, household, and individual (intercept and slope for linear time in the WHZ model) to account for the correlation at each level when estimating the variance.15
All children with complete covariate data, regardless of nutritional status, were included in the longitudinal analyses of WHZ and HAZ. The mixed effect models do not require the same number of observations on each subject, therefore children with incomplete outcome data were retained in the analysis. This resulted in an ‘all available analysis’, where all available anthropometric measurements on each subject with complete covariate data were included in the models. Observations with missing information on any covariate in the adjusted models were not included. Analyses that carry forward the last value for missing outcome data were conducted to assess the sensitivity of these results to missing data. WHZ was not calculated for children with edematous malnutrition. These observations (n=3) were therefore not included in analyses of change in WHZ or the incidence of wasting.
We estimated the intervention effect from the spline model as the difference in attained WHZ and HAZ scores between the intervention and non-intervention groups every two months and over the whole surveillance period. The overall significance of the intervention over the 8 month surveillance period was assessed using a likelihood ratio test comparing a model with main effects for linear and non-linear terms for time against one with additional interaction terms between intervention group and linear and non-linear terms for time.15 We used the likelihood ratio test to assess whether intervention effects were modified by child’s age at recruitment by comparing a main effects model against one with additional interaction terms between intervention group, time, and age. In supplemental analyses, we stratified by child’s age at recruitment.
We examined treatment effects on the incidence of wasting and stunting among children who were free from the outcome at recruitment. Mortality events included all reports where the cause for absence from study visits was reported to be death by a family member or the head of village. Children contributed person time to the analysis from recruitment until the first occurrence of the outcome, the end of eligibility when age exceeded 60 mo, or the end of study in March 2007. Incidence rates by village were estimated as the number of observed events over the number of child-months contributed to follow-up. Incidence rates by intervention group were estimated by taking the mean of the corresponding village incidence rates, weighted by the number of child-months from each village that contributed to the mean.16 We calculated incidence rate ratios by dividing the weighted mean from intervention group by the weighted mean from non-intervention group. Confidence intervals around the rate ratios were estimated using the Taylor Series approximations to obtain standard errors.17
Next, we estimated adjusted hazard ratios from a marginal Cox proportional hazards model with time from recruitment to the event (wasting, stunting, or death) as the outcome and calendar month as the time scale. We adjusted for child’s age at recruitment, sex, baseline HAZ and district. Baseline WHZ was adjusted for when wasting was the dependent variable. Confidence intervals used robust estimates of the variance to account for clustering at the village-level.
We calculated the prevalence of malaria, respiratory infection and diarrhea by village as the number of visits with a positive diagnosis divided by the total number of visits. Prevalence by intervention group was calculated by taking the mean of the corresponding village prevalences, weighted by the child-months of observation from each village. Confidence intervals around the prevalence ratios were estimated using the Taylor Series approximations to obtain standard errors.17 We estimated adjusted odds ratios from generalized linear mixed effect models with presence of the morbidity as the outcome and predictors that included intervention group, child’s age at recruitment, sex, baseline HAZ, district, and calendar month. Confidence intervals were adjusted for clustering at the village, household, and individual levels using random effects. We used the Wald test to assess whether intervention effects on morbidity were modified by child’s age at recruitment. In the analysis of all binary outcomes, observations with missing outcome data were assumed to represent the non-occurrence of an event. Additional analyses were conducted that assumed an event did occur when the outcome data is missing or censored that observation to assess the sensitivity of the results to missing data.
A value of P ≤ 0.05 was considered statistically significant. No adjustments were made for multiple comparisons. Analyses were conducted with the use of the Statistical Analyses System software (SAS Institute, Inc., Cary, NC) and MLwiN 2.0 (Institute of Education, London, United Kingdom).
The study protocol was approved by the the Government of Niger and the Comité de Protection des Personnes, “Ile-de-France XI”, France, and the study was authorized by the Ministry of Health of Niger. The Harvard School of Public Health granted an exemption for the Harvard investigator to conduct the data analyses with the previously collected data. Approval from all heads of villages was received prior to the start of the study. The objectives of the study and study protocol were explained to heads of households with children aged 6 to 60 mo before inclusion. An informed consent statement was read aloud in the local dialect before being signed or fingerprinted by the head of household or child care giver. This trial was registered with ClinicalTrials.gov, number NCT00682708.
The overall sample size was 3,533 children, corresponding to 1,407 households. Forty-five percent of children (n=1592) were between 6 to 24 mo of age at recruitment. Mean maternal age was 26.6 ± 6.7 years and educational attainment among mothers was low, with only 3% ever attending school. Socio-demographic characteristics of children at recruitment, including age, sex, ethnicity, maternal age and maternal education, and the prevalence of wasting at recruitment did not differ by intervention group (Table 1). Children in the non-intervention group were more likely to be stunted. During the 8 month surveillance period, there was a median of 8 visits per child (mean 6.9 ± 2.0). The number of children with height and weight measurements in August, October, December and February was 1,477, 1,475, 1,391 and 1,452 in the intervention group and 1,689, 1,635, 1,545, and 1,574 in the non-intervention group, respectively. Children contributed a total of 25,012 months to follow-up for the survival endpoint. Follow-up was similar in the two groups.
We found a significant difference in the rate of change in WHZ by intervention group over the 8 month surveillance period (P <0.001). The WHZ differences in the intervention and non-intervention groups at baseline and at the end of follow-up were −0.10 Z (95% CI: −0.23, 0.03) and 0.12 Z (95% CI: 0.02, 0.21), respectively; thus, the overall effect of the intervention on WHZ change over 8 mo was 0.22 Z (95% CI: 0.13, 0.30). Mean adjusted WHZ differences between the intervention and non-intervention groups were 0.21 Z (95% CI: 0.12, 0.29) in October, 0.09 Z (95% CI: 0.01, 0.18) in December, and 0.11 Z (95% CI: 0.02, 0.20) in February (Figure 2). There was no significant interaction by child age at baseline (P for interaction = 0.07). The intraclass correlation coefficient was 0.015.
The overall rate of HAZ change differed by intervention group over the 8 mo surveillance period (P < 0.001). The HAZ difference in the intervention and non-intervention groups was −0.06 Z (95% CI: −0.18, 0.06) at baseline and 0.08 Z (95% CI: −0.02, 0.18) at the end of follow-up. The effect of the intervention on HAZ change from baseline to the end of follow-up was thus 0.14 Z (95% CI: 0.11, 0.18). The difference in HAZ change between the intervention and non-intervention groups was 0.06 Z (95% CI: −0.04, 0.16) in October, 0.09 Z (95% CI: −0.01, 0.19) in December, and 0.08 Z (95% CI: −0.02, 0.18) in February (Figure 3). Results for differences in WHZ and HAZ did not appreciably change with the last value carried forward for missing outcome data.
Among children without each of these outcomes at recruitment, the absolute rate of wasting and severe wasting, respectively, was 0.17 events per child-year (140 wasting events / 841 child-years) and 0.03 events per child-year (29 wasting events / 943 child-years) in the intervention group, as compared to 0.26 events per child-year (233 severe wasting events / 895 child-years) and 0.07 events per child year ( 71 severe wasting events / 1,029 child-years) in the non-intervention group. The intervention thus resulted in a 36% (95% CI: 17% – 50%) reduction in the incidence of wasting and a 58% (95% CI: 43% – 68%) reduction in the incidence of severe wasting (Table 2). These effects were not modified by age at recruitment. There were no significant effects on the incidence of stunting. These results did not change under alternative assumptions for missing data.
The mortality rate was 0.007 deaths per child-year ( 7 deaths / 986 child-years) in the intervention group and 0.016 deaths per child-year (16 deaths / 1,099 child-years) in the non-intervention group, resulting in a non-significant 49% (95% CI: 0.25, 1.05) reduction in all-cause mortality (Table 3). We did not observe an effect of the intervention on the prevalence of malaria, diarrhea or respiratory infection.
This cluster randomized trial examined the effect of short-term, preventive supplementation with RUTF on the nutritional status, mortality, and morbidity children 6 to 60 mo of age. We found a protective effect of the intervention on WHZ change and a significant reduction in the incidence of wasting and severe wasting. There was a statistically non-significant reduction in mortality with RUTF supplementation.
This is the first population-based study to evaluate the effectiveness of RUTF in the prevention of wasting, but the protective effect of this intervention on WHZ decline and wasting incidence is consistent with the therapeutic use of RUTF in a variety of settings.3, 6, 8, 9, 18 RUTF has been shown to increase energy and micronutrient intake in children < 5 years.6, 18 The increase in energy intake associated with RUTF supplementation likely contributes to weight gain. The possibility of weight gain due to improved appetite from increased micronutrient intake has been suggested by others 19 but has not been consistent.20
This study found a mean adjusted difference in WHZ between the intervention and non-intervention groups from baseline to the end of follow-up of 0.22 Z. An increase of this magnitude in the mean WHZ will shift the population distribution of WHZ to the right and reduce the prevalence of wasting and severe wasting. The exponential relationship between the risk of mortality and nutritional status 21 suggests that the clinical importance of this intervention effect for mortality will be greatest on the left part of the WHZ distribution among children with low WHZ scores.
Sample sizes were not calculated to estimate differences in groups between the periods during and after supplementation. However, the data suggest that the intervention effect on WHZ change was greatest during the 3-month period that coincided with the actual administration of the supplement and a period of acute food insecurity preceding the harvest. Only a small benefit of supplementation appears to be sustained in the months after supplementation ceased. This suggests that short-term supplementation with RUTF may be targeted to suitably address specific, short-term nutrition needs, but further study is required to assess possible long-term improvements.
We found a limited effect of RUTF supplementation on HAZ, but the magnitude of difference in HAZ is in the range reported in trials assessing the effectiveness of complementary feeding practices in older infants.22 The small effect on HAZ change found here is likely due to the short duration of supplementation. The three-month intervention may have been too short to demonstrate an important impact on linear growth. A recent review of complementary feeding interventions suggests that the impact of similar programs on linear growth has been inconsistent, with significant improvements achieved in only some settings.23
Twenty-five children died during the study period, with less than half as many deaths in the intervention group than in the non-intervention group. A study from Malawi on the effectiveness of home-based treatment with RUTF found a similar non-significant decrease in mortality risk associated with RUTF supplementation compared to standard therapy.3 Data on the reported cause of death in this study suggest that the reduction in mortality with RUTF supplementation was achieved through protection against malnutrition (2/18 deaths in the non-intervention group vs. 0/7 deaths in the intervention group) and malaria (7/18 deaths in the non-intervention group vs. 2/7 deaths in the intervention group). Cause of death, however, was determined by verbal autopsy, which is not well-suited to distinguishing between causes of deaths with similar features and may suffer from misclassification.24 Interpretation of these data will therefore require caution.
There was no evidence of increased risk of malaria associated with RUTF supplementation. Findings of adverse health effects due to iron and folic acid supplementation in a large community-based randomized controlled trial in Zanzibar have suggested that iron supplementation proceed cautiously in settings where the prevalence of malaria and other infectious diseases is high.25 This study, however, suggests that RUTF, which is fortified with 11.5 g of iron / 100 g and other micronutrients, is unlikely to increase the prevalence of malaria. Further research is warranted to examine the impact of RUTF on the incidence of malaria.
The association of under-nutrition and increased susceptibility to infectious disease is well known,26 and evidence is accumulating on the possibly protective effect of some micronutrients, such as zinc, on diarrhea and respiratory infection.27, 28 The lack of a significant effect on diarrhea and respiratory infection in this study may be due to the non-specific nature of the diagnoses based on maternal report, the competing absorption of multiple micronutrients, such as zinc and iron, or insufficient dosages of these micronutrients in RUTF. Two studies have previously reported on the effect of RUTF on morbidity but results have been inconsistent.3, 18
There are several limitations to our study. First, the small number of clusters may have limited the benefits of randomization resulting in unmeasured confounding. Intervention groups did not significantly differ from each other for child, maternal and household characteristics, with the exception of a higher prevalence of stunting at recruitment in the intervention group. Imbalances in height-for-age were accounted for in all multivariate regression models. Consideration of the impact of differences in other specific baseline characteristics, including the frequency of hospitalization in the previous month and the prevalence of malaria, was also made. These factors were unlikely to strongly confound, or explain away, the observed differences attributed to the intervention due to their low prevalence. In multivariate regression models, the inclusion of these variables did not appreciably affect results. Potential measurement error of child’s age at recruitment and of anthropometric variables, such as length, may have resulted in residual confounding and reduced the statistical power to detect significant effects, respectively. This study was unblinded with respect to intervention assignment, however, we do not expect this to have had a differential impact on standardized anthropometric measurements. It did not appear to affect follow-up. This study also was not able to collect complete response data on all children, introducing the potential for bias if the mechanism for missingness is not ignorable. The proportion of missing data, however, is relatively small at each time point during follow-up and sensitivity analyses were used to assess the potential effect of missing response. Different strategies to account for the missing data did not appreciably change our conclusions. Finally, we were unable to measure dietary intakes at recruitment or during the intervention. We therefore did not have information on average energy intake, the macro- and micronutrient composition of baseline diets or information on energy received from the supplement vs. household foods during the intervention to indicate whether RUTF supplemented or displaced usual intake. Compliance was similarly not measured, limiting our understanding of how the supplement was used by each child and within the household.
The likelihood of contamination was reduced using village- rather than individual-level randomization. Contamination between intervention and non-intervention villages is also unlikely due to their geographic separation. There was no evidence of re-sale of the supplement in local markets to suggest that individuals from non-intervention villages would have been able to access the supplement provided in this study outside of the study. No secular changes in the health and nutritional status of children in the study villages were observed during the 8 mo of follow-up.
These results are applicable to other settings of acute food insecurity, where access to food is limited due to emergency or seasonal conditions, and short-term food supplementation is required for the prevention of wasting. The effectiveness of preventive supplementation with RUTF in other settings may depend on RUTF acceptability, the extent of re-sale after distribution, and the adequacy of the public health and nutrition systems in place. Further research is warranted to identify the minimal dose required to achieve an effect and to compare the impact of other formulations of RUTF and locally available diets, which also may be effective in improving nutritional status in children.29, 30 Information is also needed on the cost-effectiveness and feasibility of large-scale RUTF distribution. The relatively high costs of imported RUTF (USD 4.54 / kg before duties and shipping, written communication Guillaume Sauvage, Médecins Sans Frontières, Paris, France, July 2008) and locally produced RUTF (USD 3.66 / kg before duties, written communication Mark Manary, Department of Pediatrics, Washington University, School of Medicine, St Louis MO, USA, July 2008) may challenge the effective scaling up of short-term experiences such as these.
In conclusion, this study demonstrates that the distribution of RUTF to non-malnourished children 6 to 60 mo of age can be effective in limiting reductions in weight-for-height and reducing the incidence of wasting and severe wasting in the short-term. The effectiveness of any intervention to prevent malnutrition, however, will ultimately depend on its consideration of the underlying causes of malnutrition, integration with other broad-based strategies to improve public nutrition, and feasibility within the resource constraints of humanitarian and public health programming.
We thank the Ministry of Health of Niger, in particular Amina Yaya, MD (Nutrition Division) and the Regional Public Health Office of Maradi for their support of this project. We thank the field teams of Epicentre and Médecins Sans Frontières, our dedicated teams of translators, research nurses and nutritional assistants for their support in gathering data. In particular, we wish to thank the Program Director, Isabelle Defourny, MD (Médecins Sans Frontières), for her critical and precious insight into the operation of the MSF program in Niger and support of this research and Olivier Cheminat, MSc (Epicentre), Thomas Roederer, MSc (Epicentre), Nael Lapidus, MD (Epicentre), Emmanuelle Robert, MA (Epicentre) and Alexandra Simon, RN (Epicentre) for their dedication and work on ensuring the data collection in this study. We also wish to thank the Head of Mission, Thierry Climat, MA (Médecins Sans Frontières), Medical Coordinator, Susan Shepherd, MD (Médecins Sans Frontières), and Field Coordinator, Gwenola Seroux, RN (Médecins Sans Frontières), as well as Emmanuel Drouhin, BA (Médecins Sans Frontières)for their help in facilitating this study. André Briend, MD, PhD (World Health Organization) provided important comments on both the protocol and draft of this manuscript. We also sincerely thank Alain Moren, MD, MPH, PhD (Epiconcept), Donna Spiegelman, ScD (Harvard School of Public Health), and P. Gregg Grennough, MD,MPH (Harvard Humanitarian Initiative) for their statistical advice and comments. No individual has received compensation for this study.
Funding / support
This study was supported by Médecins sans Frontières (MSF). Sheila Isanaka was supported in part by the National Cancer Institute (grant R25-CA098566).
Trial Registration This trial was registered with ClinicalTrials.gov as “Sentinel Surveillance of Acute Malnutrition in the Region of Maradi, Niger, Children Under 5: Impact of a Preventive Intervention,” number NCT00682708.
Author ContributionsIsanaka had full access to all of the data in the study and takes responsibility for the integrity of the data and the accuracy of the data analysis. Study concept and design: Nombela, Poupard, Van Beckhoven, Gaboulaud, Guerin and Grais. Analysis and interpretation of the data: Isanaka, Nombela, Djibo, Guerin and Grais. Drafting of the manuscript: Isanaka. Critical revision of the manuscript for important intellectual content: Isanaka, Nombela, Djibo, Poupard, Van Beckhoven, Gaboulaud, Guerin, and Grais. Statistical expertise: Isanaka and Grais. Obtained funding: Guerin and Grais.
Conflict of interest
Role of the sponsor
MSF reviewed the final study protocol as described here and had no role in the design and conduct of the study, collection, management, analysis and interpretation of the data or preparation of the manuscript.