Adolescent childbearing rates are higher in Central America than almost anywhere else. However, in this research we discovered that adolescent childbearing exhibits variability from one village to another, and we might discover factors associated with this spatial variability that can help us understand key characteristics underlying the pattern of early childbearing. To do this, we assessed the village-level normative and network factors associated with adolescent birth (birth taking place before age 20 years) in rural Honduras and evaluated the geographic dispersion of these patterns. We used full population data from 24,937 people in 176 villages (81% of the eligible population) to assess prevalence and patterns of adolescent childbearing among women. We modeled the predictors of adolescent births among women younger than 21 years. After accounting for individual demographic characteristics, one of the strongest predictors of adolescent birth within the population was village-level collective norms about the acceptability of adolescent childbearing, based on aggregating normative measures from the entire population. The proportion of women in the village who had given birth as an adolescent was also strongly associated with an individual girl's likelihood of having given birth as an adolescent. We used full village-level network analyses to calculate social cohesion within the village. Normative pressure was strongly associated with the likelihood of an adolescent birth in villages with high cohesion (high network density) and was not associated or had a weak association in villages with low cohesion. On the other hand, the longer a girl had lived in the village, the stronger the association between the overall proportion of women in that village who gave birth as adolescents and the girl's own likelihood of having done so. Spatial analyses suggest that levels of adolescent births vary spatially across villages, as do the village-level normative factors associated with them. •Honduras has very high rates of adolescent fertility, especially among rural poor.•We use data from 176 villages in Honduras  including social networks.•Village norms  are associated with adolescent fertility.•The impact of norms varies by time in village, and village level social cohesion.•There are geographic groups of villages with high norms and high adolescent fertility. Within Once adolescent girls become mothers, disadvantage can become enhanced or entrenched and is often passed on to their own daughters. This cyclical process is inextricably linked to a long list of individual risk factors, most of which are associated with a life of disadvantage. Determinants of adolescent childbearing in Latin America include living in poverty , experieAlthough individual risk factors can be strongly predictive of adolescent birth, higher-level interpersonal and community factors are an important part of the equation, albeit one that is studied less frequently. Previous research has shown that the risk of adolescent childbearing within Latin America decreases within communities and families that are socially cohesive, that emphasize respect for family, and that hold strong norms against early pregnancy . Girls w22.1diffusion of innovations has mostly focused on descriptive norms, which operate through a mechanism of social learning  and perhaps more importantly because it is through y occurs , 2015. In common . For exan common . Spatialby place  For inst3Here we utilize a unique dataset with a complete census of the population from 176 villages in rural Honduras, their social network connections, and spatial data to assess whether village-level normative and social network factors are associated with giving birth before age 18 years, with a focus on delineating the role of individual-level normative factors from those held at the community or village level. Spatial data allow us to determine whether there are spatial patterns of adolescent births and the norms associated with them, an important consideration for programs designed to decrease rates of adolescent childbearing. We hypothesize: 1) in communities in which there are injunctive norms against adolescent childbearing, adolescent girls may be less likely to become mothers; 2) in communities in which a high proportion of women gave birth as adolescents (strong descriptive norms), adolescent girls are more likely to become parents; 3) adolescent girls in more socially cohesive communities are less likely to have given birth as adolescents; 4) normative influence is stronger in more cohesive communities; and 5) these normative patterns are geographically clustered.44.1Our study uses full population census data from the western municipalities of the largely rural Copán department of Honduras see  to analyWe completed geographical mapping for 176 villages chosen for the randomized trial located in the municipalities of Copán Ruinas, Santa Rita, Cabañas, and San Jerónimo in the department of Copán, allowing us to gain more precise calculations of the study population and field conditions, including terrain, rainfall, and distances to health facilities. The area is over 200 square miles of rugged mountainous terrain with an estimated total population (in the 176 villages) of 32,800 people older than 12 years . We conducted a census in 2016 with 92% of the eligible population, of which 25,032 completed a baseline survey that included sociocentric and behavioral health measures. For the purposes of the fertility analyses, we excluded children younger than 15 years (N = 2577), as they did not complete the full reproductive history and the proportion who had already experienced a birth was miniscule. Individuals who were cognitively impaired and unable to provide consent were also excluded (30). Our final number of participants was 25,032, of which we used data from 22,449 for the fertility analyses reported in this paper.4.2http://humannaturelab.net/resources/software/trellis/) to undertake the main survey, which included a battery of “name generator” questions to capture social relationships. In this study, the boundaries of each network were the villages, so that individuals could nominate any individual from within their own village as a social contact. Photographs were taken of all persons from whom data were collected, and they were used to validate the social contacts named by the respondents.We used the publicly available Trellis software . The outcome, adolescent birth, was defined in two ways. Our definition of adolescent motherhood was having had a child before age 20 years, which is consistent with the definition used in DHS and other similar demographic surveys.4.3.1Female respondents were asked whether or not they had ever given birth to a living child. Women who reported having given birth were then asked to provide the birthdates of their last four children. For women with four or fewer children, their age at first birth was calculated as the difference between their date of birth and the date of birth of their first child. For women with more than four children, the age at which they first gave birth was approximated using a question that asked the age at first pregnancy. None of the women in our primary sample population of women aged 15–20 years had more than four children.4.3.2Individual demographic controls included age, marital status, religion, income sufficiency, education, food security, proportion of life lived in the village, and indigenous status. For exact coding of these measures please see the online appendix.4.3.3All respondents were asked their personal attitudes regarding the appropriate age for a first birth for women: “At what age is it OK for a girl to have her first baby?” We also asked each respondent about injunctive norms around adolescent birth: “If a girl younger than 18 has a baby, will people in the community think this is good, bad, or neither?” We modeled normative beliefs in support of adolescent birth as a binary variable, “good” or “bad/neither,” as the statistical model showed no difference in the association between “bad” or “neither” with adolescent birth but a strong difference between “good” and “bad/neither.” In this case, coding the variable as continuous would have resulted in an artificial result suggesting a linear relationship.4.3.4entire sample of men and women of all ages for each village to get a comprehensive understanding of the village-level social environment.Because the unit of analysis is women younger than 21 years, all individual-level variables are specific to that population. However, we constructed the village-level normative and social network factors using the 4.3.5We aggregated the means of individual attitudes regarding the appropriate age for first birth (collective attitudes), and the proportion of each village that reported perceptions of norms regarding giving birth before age 18 years as “good” (collective injunctive norms). As a proxy for descriptive norms, which would ideally be measured by asking each respondent what they think is normally practiced, we calculated the proportion of women in the village overall who gave birth before age 20 years (descriptive norms).4.3.6density, a measure of cohesion at the network level, using the comprehensive network constructed from the relationships across all relationship questions and across all members of each village. Density is a measure of the number of identified ties over the total number of possible ties  for the approximate geographic center of each village in the dataset. .4.4As noted above, our primary unit of analysis for our predictive models was individual women between the ages of 15 and 20 years at the time of the survey (N = 2990), with the primary outcome being adolescent birth, or having given birth before age 20 years. All of our analyses were based on multilevel logistic regression, clustering at the village level. A −2 log-likelihood test confirmed significant village-level clustering and the appropriateness of using a multilevel model. Continuous variables, including village-level network measures and proportions, were z-score centered. Our initial models included individual-level factors, and subsequent models included village-level social normative factors controlling for village size. Finally, we ran interactions testing whether village-level normative factors were moderated by village-level cohesion, to determine whether or not cohesion alters the impact of significant village-level normative factors on our outcomes.4.4.1Despite the overall high level of adolescent childbearing in this rural area of Honduras, there is considerable variability both within and between villages in the likelihood of adolescent birth. Although statistically significant village-level clustering is important evidence that supports the validity of our findings on the association between village norms and the level of adolescent births, it does not answer the question of whether this village-level variation is spatially dependent. In other words, are there clusters of villages with higher or lower rates of adolescent fertility, with associated higher or lower rates of norms in favor of it?We used the Getis-Ord Gi* hot spot statistical analysis tool to determine the existence of statistically significant spatial clusters of high or low values. As social norms tend to cluster within geographic areas, we conducted a series of spatial analyses to discern to what extent the incidence of adolescent birth and the norms associated with it are spatially significant. Our spatial analyses included hot spot analysis using the Getis-Ord Gi* statistic (within ArcGIS) to examine significant spatial clustering of villages with regard to norms surrounding approval of adolescent birth, as well as the recent incidence of giving birth before age 18 years. We further explored the spatial patterns in our analysis to evaluate the presence of spatial non-stationarity—spatial variation in the regression relationships. In other words, do we find evidence that there are geographic differences in the strength of the association between village-level normative beliefs and the rate of adolescent childbearing in these villages? We used geographically weighted regression to assess possible spatial differences in the association between demographic and normative factors and adolescent birth within the village. We also used the ArcGIS spatial clustering grouping analysis tool, with spatial constraints set to K-nearest neighbors (using a minimum of 8 neighbors), to identify spatially contiguous villages that have similar characteristics with respect to levels of and norms regarding adolescent birth. These could be thought of as “neighborhoods” of similar patterns of adolescent fertility. Note that this is an exploratory tool, but the results can be very suggestive of underlying social processes even if not definitive.55.1Adolescent birth is common within this population of women. Across the entire sample of women, including those younger than 21 years, 44% had given birth as an adolescent see . Among t5.2Our first set of analyses focused on individual girls and women between the ages of 15 and 20 years (N = 2990). 5.3Testing bivariate and multivariate associations at the village level  for every 1 SD increase in both the proportion of the village that believes the community supports adolescent birth and the village-level proportion of women who gave birth as adolescents. These effect sizes are approximately one-third of what we found for a 1 SD increase in education, which is one of the most well-documented factors associated with adolescent childbearing. When we set model parameters to their means, we find that the predicted probability of a girl having an adolescent birth is 20%  when only 10% of the village believes that the community supports giving birth before age 18 years, compared with a probability of 32%  when 40% of the village believes that the community supports giving birth before age 18 years.5.5P = .01). Stratifying by cohesion illustrated a lack of association between the proportion of the village who believe the community supports adolescent birth and adolescent birth for low-cohesion villages, and a strong association for high-cohesion villages. Setting all parameters equal, when 40% of the village report that the community supports adolescent birth, the probability of adolescent birth is 35%  in high-density villages versus 17%  in low-density villages. Village-level social network cohesion (network density) did not independently predict the risk of an individual woman having an adolescent birth. However, in an interaction model , Model 3We did not find a significant interaction between cohesion and the proportion who had given birth as an adolescent. However, we found significant interactions between the proportion of life a spent in the village, the proportion of women in the community who had given birth as an adolescent, and a girl's likelihood of having given birth as an adolescent. The greater proportion of a girl's life spent in the village, the stronger the association between the descriptive norms, the proportion of women who had given birth as an adolescent, and the likelihood of a girl having given birth as an adolescent. Both of the above mentioned interactions together (not shown) retained significance in a final model, with an AIC of 2606, a 13-point decrease from Model 2, including no interactions. This is evidence that both interactions in the model improve model fit.5.5.1Some of the individual-level controls we used for this analysis, although potentially important confounders for the relationship between village-level factors and adolescent birth, could be the result of having given birth as an adolescent. Therefore, we ran our main model, Model 2, eliminating education, income, food security, time spent in the village, the individual injunctive norms, and the individual attitudes measures . Our res5.6A high value (hot spot) indicates that a village is surrounded by other villages with high values, whereas a low value (cold spot) indicates that a village is surrounded by other villages with low values. Using the proportion of women between the ages of 15 and 20 years who gave birth as adolescents as our outcome variable and the proportion of village respondents who report positive community norms regarding adolescent birth as the main predictor, we found significant geographic variation in the statistical association of normative beliefs and adolescent birth at the village level. Rather, the relationship was consistently strong throughout the department. Generally, where there are hot and cold spots of norms in support of adolescent childbirth, the association between village-level norms and village-level childbirth is the highest see .Fig. 5GeThe next step in the spatial analysis was to adjust for the fact that hot spot analyses assess the spatial clustering of single variables but cannot incorporate multiple variables and their associations. Therefore, we used ArcGIS to run a grouping analysis—a type of cluster analysis that takes the spatial distribution of multiple variables into account. This allowed us to create “neighborhoods” of villages exhibiting similar levels of both adolescent birth and normative beliefs in support of adolescent birth. The Calinski-Harabasz pseudo F-statistic, which is a ratio reflecting within-group similarity and between-group difference, suggested that four groups would be optimal.6In this analysis, we explored an unusual data set from rural Honduras that included demographic, behavioral, attitudinal, normative, and social network measures for a full population from 176 villages to consider the village, network, and normative predictors of adolescent birth among women between the ages of 15 and 20 years at the time of the survey. We predicted that village-level social network characteristics and village-level normative characteristics would be independently associated with the likelihood that a girl gave birth as an adolescent, and that these associations would be geographically clustered. Our hypotheses were partly borne out, in ways that can provide important insights for intervention efforts to reduce adolescent fertility in similar communities in Central America.We expected to find that social cohesion, as measured by social network density, would be significantly associated with a decreased risk of adolescent birth, by virtue of implied higher levels of social surveillance. Because adolescent birth is associated with social instability and lower levels of social control , more-coThe fact that network density changes the association between collective norms and adolescent births is strong evidence that there may be causal connections in these dynamics. Injunctive norms work through direct social influence. The stronger the network, the stronger the influence. Highly dense networks reinforce and transmit norms much more than loosely connected networks . AlthougConsistent with the consideration that village-level normative factors may be an important factor underlying adolescent births in these contexts, our hot spot analysis and spatial grouping analyses showed significant geographic clustering of adolescent births and village-level norms—both hot spots and cold spots. Particularly noteworthy is the strong geographic cluster of high normative and high fertility villages in the eastern edge of our study region. This is important evidence that although norms and their associations are occurring at the village level, these associations are also occurring in geographic space. The fact that our geographically weighted regression showed spatial differences in the associations is also telling. Normative influence seems to be occurring where there are hot spots in norms and hot spots in adolescent childbearing rates. Understanding these contextual differences can be a crucial element to successful interventions to prevent adolescent childbearing in similar settings. Geographic clustering of norms and adolescent childbearing suggest that a blanket approach across departments could be ineffective. Understanding where there are clusters of risk both geographically and socially, and how to identify those clusters, can help interventionists target them far more effectively.There are limitations to these analyses. First, the question about age at first birth is only asked of women who reported a live birth, so our estimates are likely to underestimate the earliest ages at which girls are becoming pregnant. Because these are cross-sectional data, we cannot track time-dependent associations between our predictors of interest and adolescent births. All questions are based on self-report so there is the possibility of response bias in some of our measures. It is also important to recognize that women were asked about their attitudes (and their view of community attitudes) after their first birth, so it is impossible to disentangle cause and effect. For some women, giving birth as an adolescent may be consistent with their prechildbearing views, whereas other women may be justifying their behavior after the fact. It is possible that our findings could be skewed by omitted variable bias. Village-level clustering of adolescent childbearing and its association with social norms may be driven by village-level factors for which we have no measures. Our future research will address these issues, as well as take a deeper look at the individual social dynamics through individual network analysis.Despite these limitations, these data provide a rare opportunity to analyze a full census of a population, including a detailed reproductive history. The magnitude and consistency of our results suggest that adolescent fertility among girls in these populations is not only a common and serious issue but also depends on collective pressures organized in geographic and social space. Individual-level risk factors of adolescent fertility are fairly well established, and, in our analyses, we find individual-level results consistent with previous research. However, after controlling for these individual attributes, we still find variation in adolescent fertility rates. What does that variation come from? Our results show it is a combination of social normative and social network effects. Although social normative effects are the most strongly associated and are clustered geographically, village-level network patterns play an important role in moderating normative effects. In sum, individual attributes, geography, social norms, and social interactions all play roles in explaining variation in adolescent fertility in rural Honduras.The Yale IRB and the Honduran Ministry of Health approved all data collection procedures (Protocol number 1506016012) and all participants provided informed consent before enrolment.Bill and Melinda Gates Foundation, grant number OPP1098684, and NICHD grant number K01HD087551.Funding provided by the The authors have no conflicts of interest or financial disclosures to report.


Ethiopia has one of the lowest rates of facility delivery and is promoting birth preparedness among pregnant women through its community health services to increase the rate of institutional delivery and reduce maternal mortality. Observational studies of birth preparedness in Ethiopia have thus far only reported the marginal effect of birth preparedness when controlling for other factors, such as parity and education.In this cross-sectional study, we use propensity score modeling to estimate the average population-level effect of birth preparedness on the likelihood of delivering at a facility.We conducted secondary analysis of household survey data collected from 215 women with a recent live birth within the catchment areas of 10 semi-urban health centers. A mother was considered well prepared for birth if she reported completing four of the following six actions: identified a skilled provider, identified an institution, saved money, identified transport, prepared clean delivery materials, and prepared food. We performed unadjusted and multivariate logistic regression analyses, with and without propensity score weighting, to assess the relationship between birth preparedness and institutional delivery.One hundred respondents (47%) delivered in an institution, and over two-thirds  were considered well prepared for birth. Institutional delivery was more common among women who were considered well prepared (57%) versus those who were considered not well prepared (19%). In the model with propensity score weighting, women who were well prepared for birth had 3.83 times higher odds of delivering at a facility .This study contributes to existing evidence supporting the inclusion of antenatal birth preparedness counseling as a part of an antenatal care package for promoting institutional delivery. Important gaps remain in operationalizing the definition of birth preparedness and understanding the pathway from exposure to outcome. Globally, over 300,000 mothers die annually due to pregnancy- and childbirth-related complications. Ninety-nine percent of these deaths occur in developing countries, with Sub-Saharan Africa absorbing over half (66%) of this burden . The majOne of the strategies to increase the coverage of skilled attendance at birth is the promotion of birth preparedness among pregnant women. The “three delays” model describes three junctures at which delayed care can result in maternal deaths: delay in deciding to seek medical care, delay in reaching medical care, and delay in receiving care . Birth p9marginal effect of birth preparedness beyond other factors influencing facility delivery that are included in the statistical model [Ethiopia, an East African country with one of the lowest global coverage rates of skilled attendance at delivery 27%) , has inc141516171819202122232526% , has i28al model . Given tFor this study we conducted secondary analysis of household survey data collected for an evaluation of community-based promotion of skin-to-skin care (SSC) and exclusive breastfeeding by Health Extension Workers (HEWs) in Ethiopia, the results of which are published elsewhere . While pThe household survey, for which detailed methods have been published 48, took The primary outcome variable for this analysis is institutional delivery, which includes delivery in a government hospital, government health center, nongovernmental organizational clinic, or private hospital. The main predictor variable is a composite indicator for completing birth preparedness actions. Researchers and practitioners have varying interpretations of the knowledge, intentions, and behavioral elements that constitute birth preparedness and with which combination of these a woman is considered well prepared for birth. We narrowly defined birth preparedness in terms of specific actions or behavioral practices the mothers reported undertaking during pregnancy. Following a similar approach as that in other studies 151622232632,222326Our definition of birth preparedness was developed through a review of 17 peer-reviewed papers on birth preparedness in Ethiopia and similar settings , a review of the HEW program guidelines on birth preparedness counseling , and con28Sociodemographic and health care utilization variables that we considered as potential confounders were maternal age, marital status, parity, educational level, household wealth level, knowledge of pregnancy danger signs, distance to a health center, and time of labor start. We used principal component analysis to produce a wealth score for each household using housing material, toilet source, drinking water source, and household assets. A woman was considered knowledgeable about pregnancy risks if she could name at least one of the five danger signs without being prompted : fever, We included all 215 respondents in the dataset to analyze the effect of birth preparedness on institutional delivery. Descriptive statistics for the prevalence of birth preparedness actions were calculated adjusting for survey design. In order to estimate a population-level effect of birth preparedness on facility delivery—rather than the marginal effect estimated by multivariate regression models—we used propensity scoring methods to improve the comparability of the treatment groups on characteristics we could measure. This propensity score analysis uses the weighting approach, which provided better balance between treatment groups than a full matching approach  and was Sociodemographic characteristics. Table Use of reproductive health services and other obstetric characteristics. One hundred (47%) respondents gave birth in an institution. About one-third of the respondents  were primipara mothers, half had given birth to 2–4 children , and the remaining 41 respondents (20%) had five or more children. The respondents were about evenly divided between those who had or had not received four or more ANC visits by the time of labor start. Half of the respondents lived more than one hour from the nearest health center . The great majority  of the respondents knew at least one danger sign. The results of the unweighted and unadjusted logistic regression . The results of Model 1 with sample weights applied were OR = 4.22, 95% CI: 1.61–11.05, p-value = 0.005.Model 2. An extension of Model 1, we controlled for potential confounding using propensity score weighting for the unadjusted outcome. The results suggest that women who were well prepared were at least three times more likely to have given birth in an institution  compared with women who were not well prepared. When we applied a composite weight (propensity score multiplied by sample weight), the results were OR = 3.44, 95% CI: 1.18–10.02, p-value = 0.025.Model 3. This model assessed the association between institutional delivery and birth preparedness and adjusted for other potential confounding covariates using multivariate logistic regression without propensity score weighting. Women who were well prepared were at least three times more likely  to deliver in an institution compared with women who were not well prepared. The results of Model 3 with sample weights applied were OR = 3.06, 95% CI: 1.11–8.39, p-value = 0.031.Model 4. An extension of Model 3, Model 4 looked at the association between institutional delivery and birth preparedness using a multivariate logistic regression model with propensity score weighting. The results suggest that well-prepared women were at least four times more likely to have delivered in an institution  compared with women who were not well prepared. When we applied a composite weight of propensity score multiplied by sample weight, the results were OR = 4.84, 95% CI: 1.62–14.51, p-value = 0.006.This study provides what we believe is the first estimate of the average population-level effect of birth preparedness on facility delivery in Ethiopia. Across models, well-prepared women were at least three times more likely to give birth in a facility compared with women who were not well prepared. In fact, the estimated average population-level treatment effect of birth preparedness reported in the propensity score models (2 and 4) was similar in direction and magnitude to the marginal effect of birth preparedness reported in the standard multivariate regression models (1 and 3). In bivariate analyses, planning to deliver in a health facility and with a skilled attendant were the two birth preparedness components with the strongest association with facility delivery among the six that we measured.This study also contributes to the rapidly growing literature on the prevalence of birth preparedness actions among pregnant women in Ethiopia. A large proportion of respondents (71%), who were drawn from a semi-urban population with health indicators generally higher than women in a primarily rural sample , were we151620212223242830While we found a positive association between birth preparedness and institutional delivery, previous research on the effectiveness of birth preparedness promotion and care-seeking has been mixed. Studies in Ethiopia, Tanzania, and Nepal reported positive associations between birth preparedness and institutional delivery. Two unmatched case-control studies in Ethiopia reported positive marginal effects of birth preparedness on facility delivery: one study reported a six times greater odds for institutional delivery (adjusted OR (AOR) = 6.96, 95% CI: 2.42–19.99, p-value not provided) ; and ano842Among the studies from Ethiopia, there are a number of potential reasons for the variations in the rate of birth preparedness and its association with institutional delivery. First, differences in the make-up of the samples, with regard to variables reported to be predictors of birth preparedness, could be an influence. Residence, educational status, ANC attendance, parity, wealth level, knowledge of pregnancy danger signs, and past obstetric complications are some variables associated with an increased likelihood for preparing for birth 1415119202123.192021In Ethiopia and elsewhere, an important consideration for interpreting the results of birth preparedness studies is the varying definitions of birth preparedness used by each study, both in terms of what elements are included and how many actions a woman must complete to be considered well prepared. While the concept of birth preparedness encompasses knowledge, intentions, and behaviors, researchers have differed in their operational definitions and measurements of birth preparedness by focusing on select elements within each, some, or all of the components. As was done in this present study, some researchers treat knowledge as a preceding step to birth preparedness actions; thus, they do not include knowledge within the primary independent variable (birth preparedness) 15182144.1518218The definitional inconsistencies of birth preparedness prevent a clear understanding of the effectiveness of birth preparedness counseling and what specific actions lead to institutional delivery across studies and settings. Practitioners looking to translate research into public health programs would be rightfully confused when trying to determine which elements of birth preparedness have been shown to be effective and incorporate those elements into programs. Brazier et al. disassembled the concepts of birth preparedness knowledge, complication readiness knowledge, and birth preparedness actions, to create separate composite variables for the components of these two constructs . In theiThis study compared results, with and without sample weights, for propensity score weighting (Model 2) and regression-adjustment (Model 3), and we combined propensity score weighting with the regression-adjustment approach (Model 4) to control for error when estimating the association between birth preparedness and institutional delivery. Although all models in our analysis consistently found a positive association between birth preparedness and facility delivery, we observed extreme survey sample weights in the control group (birth preparedness = no). When propensity score weighting was combined with the composite weight (propensity score multiplied by the sample weight), this imbalance was further exaggerated, which resulted in additional variability in the measure of association, as seen in the results of Model 4. This is one of the known issues to propensity score weighting . While pOur study has several limitations. First, the study design is cross-sectional and relies on a woman’s retrospective, self-reported preparations for childbirth. Self-reports may be subject to increased recall bias as time elapses. However, the study participants were questioned within 7 months of giving of birth, which is a shorter time period than what has been reported  in many observational birth preparedness studies. Second, women’s retrospective, self-reported preparations for delivery may be influenced by the care they receive during childbirth . Third, The results from this study suggest that increasing birth preparedness behaviors during pregnancy will likely increase population-level coverage of facility delivery. These findings are supportive of the Ethiopian Ministry of Health’s efforts to promote birth preparedness as a key component of ANC provided by health care workers and the work to expand coverage of ANC so all women can receive and act on birth preparedness messaging. In this study, the elements included as a part of the composite variable for birth preparedness associated with facility delivery are identifying a skilled provider, identifying a health institution for birth and making arrangements to save money, identifying transport, identifying delivery materials, and preparing food.The additional file for this article can be found as follows:10.5334/aogh.920.s1Supplemental Table 1.Definitions of Birth Preparedness in Previous Studies.


We report a case of fetal lingual cyst that was diagnosed prenatally using 2-dimensional and 3-dimensional images during routine second trimester screening. To the best of our knowledge, this is the first description of a cystic lesion disappearing before birth.A 36-year-old woman at 22 weeks’ gestation showed the presence of an oral cystic lesion in routine second trimester ultrasound screening. The lesion measured 18 × 15 × 15 mm.A follow-up ultrasound examination was performed every 4 to 6 weeks.The cyst disappeared in a follow-up ultrasound examination at 35 and 37 weeks of gestation. A male newborn who weighed 3480 g was delivered with no feeding difficulties. The boy was followed to 6 years after birth. The child had normal growth and development, and there was no recurrence of the cyst.Prenatally diagnosed lingual cysts are uncommon findings that can include many different pathologies with a wide spectrum of severity. Lingual cysts usually have a good prognosis. Additionally, a follow-up examination showed that the cystic lesion disappeared before birth. Clinical findings, sonographic features, differential diagnosis, further investigations, intrapartum management, postnatal complications, and histopathological characterization of this rare oral cystic mass are discussed and the recent literature is reviewed.Lingual cysts are uncommon findings that can include many different pathologies with a wide spectrum of severity. Prenatal diagnosis of a lingual cyst has rarely been reported.2A 36-year-old woman, G2P1, with an uneventful pregnancy, presented to Peking Union Medical College Hospital at 22 weeks’ gestation for routine second trimester screening. An ultrasound examination showed the presence of an oral cystic lesion, which measured 18 × 15 × 15 mm. The lesion was posterior to the mandible and attached to the inferior aspect of the tongue in a male fetus Fig. . HoweverA male newborn who weighed 3480 g was delivered vaginally, with Apgar scores of 10 and 10 at 1 and 5 minutes, respectively. The neonate had no feeding or breathing difficulty. No major oral cyst was identified in the tongue at a clinical and instrumental examination. We followed the child to 6 years after birth. The child had normal growth and development, with no recurrence of the cyst. The study was approved by the Institute Research Ethic Committee of Peking Union Medical College Hospital. The patient has provided informed consent for publication of the case.3 including a broad spectrum of disease, such as a ranula,–5 foregut duplication cyst,–8 cystic epignathus, thyoglossal duct cyst, and lymphangioma. Most of these lesions are benign. Ranulas have an incidence of 0.74%, and are usually secondary to a mucous leak following disruption of the sublingual salivary gland elements. Ranulas can also occur as a retention cyst derived from proximal expansion of a blocked duct.–5 A foregut duplication cyst is a benign developmental anomaly that occurs along the foregut-derived alimentary tract with an incidence of 0.02%. This type of cyst most commonly arises in the thorax and abdomen.,8 Foregut duplication cysts represent one third of gastrointestinal duplication cysts. Only 0.3% of these cysts arise in the tongue, typically from the anterior two thirds of the tongue. The cysts contain epithelial lining of ciliated respiratory, gastric, and squamous epithelium. They may originate in a defect of migration from the islands of endoderm of the primordial stomach in the 4-week embryo. Epignathus is a benign teratoma that is estimated to affect between 1 in 35,000 and 1 in 200,000 live births, and it accounts for 2% to 9% of all teratomas. Epignathus mostly involves the posterior nasopharynx, hard palate, or sphenoid bone. Occasionally, these lesions may have an intracerebral extension. These lesions mal also be associated with other malformations, including polyhydramnios, cleft palate, an abnormal mandibular structure, a bifid tongue or nose, branchial cleft cyst, and congenital heart disease. Epignathus may have cystic and solid components, and it commonly protrudes out of the fetal mouth. Thyroglossal duct cyst is a congenital developmental malformation that occurs because of incomplete obliteration of the thyroglossal duct with a general prevalence of 7%. This cyst type can be found along the path of the thyroglossal duct from the foramen cecum to the pyramidal lobe of the thyroid. Differential diagnosis among these entities might be difficult because most cystic lesions show a similar anechoic avascular ultrasound pattern. Three-dimensional techniques can help to provide potentially useful information on the size, location, extension, and composition of such masses. Definitive diagnosis can still only be made on the basis of histopathology of the lesion.Congenital lingual cystic lesions are uncommon pathological entities, reported a ranula that spontaneously resolved at 5 weeks of age. However, a cystic lesion that disappeared before birth has not been reported. In some of the reported cases, a considerable increase in lesion size required an intervention to relieve obstruction.,9,15 Pires et al and George conducted ultrasound-guided aspiration of the cyst at 26 weeks of gestation; Deloison performed cyst puncture immediately prior to delivery. Prenatal decompression of the cyst can prevent postnatal airway obstruction, especially if carried out just before birth, which can also avoid the risk of hemorrhage from the cyst. Two of the reviewed cases underwent an extrauterine intrapartum treatment procedure with aspiration of the cyst to secure the airway of the infant.,17 This technique allows maintenance of uteroplacental gas exchange in the newborn. The definitive treatment for these tumors is postnatal surgical resection.A search of PubMed showed only 18 cases of lingual cysts that were diagnosed prenatally. These cases included six ranulas, seven foregut duplication cysts, 2 thyroidglossal duct cysts, one lymphangioma, and one cystic epignathus. Prenatal recognition of such a tumor should prompt a thorough evaluation of the likelihood of airway tract and esophageal obstruction. CDFI and magnetic resonance imaging (MRI) provide the ability to visualize a fluid-filled oral cavity, oropharynx, nasopharynx, and trachea. Real-time ultrasound and MRI can demonstrate motion of the mass during swallowing. Normal amniotic fluid volume and visualization of stomach bubbles are helpful for determining if there is no distortion or compression of the esophagus. Routine follow-up should be scheduled bi-weekly to monitor growth of the lesion. Steelman et al4We report a case of a relatively small cyst with spontaneous in utero resolution. A lingual cyst usually has a good prognosis. When a lingual cyst is prenatally diagnosed, the patency of the airway tract and esophagus should be evaluated. Prenatal aspiration of the cyst or the extrauterine intrapartum treatment procedure facilitates a successful outcome for large lingual cysts.Investigation: Hua Meng, Xining Wu.Methodology: Hua Meng.Software: Zihan Niu.Writing – original draft: Yixiu Zhang, Hua Meng.Writing – review & editing: Yixiu Zhang, Yunshu Ouyang, Hua Meng.


Systematic reviews of randomised controlled trials (RCTs) have suggested that maternal vitamin D (25[OH]D) and calcium supplementation increase birth weight. However, limitations of many trials were highlighted in the reviews. Our aim was to combine genetic and RCT data to estimate causal effects of these two maternal traits on offspring birth weight.N = 122,123 for 25[OH]D and N = 61,275 for calcium). Associations between these maternal genetic variants and offspring birth weight were calculated in the UK Biobank (UKB) . We used data on mother–child pairs from two United Kingdom birth cohorts  in sensitivity analyses to check whether results were influenced by fetal genotype, which is correlated with the maternal genotype (r ≈ 0.5). Further sensitivity analyses to test the reliability of the results included MR-Egger, weighted-median estimator, ‘leave-one-out’, and multivariable MR analyses. We triangulated MR results with those from RCTs, in which we used randomisation to supplementation with vitamin D  and calcium  as an instrumental variable to determine the effects of 25(OH)D and calcium on birth weight. In the main MR analysis, there was no strong evidence of an effect of maternal 25(OH)D on birth weight . The effect estimate was consistent across our MR sensitivity analyses. Instrumental variable analyses applied to RCTs suggested a weak positive causal effect , but this result may be exaggerated because of risk of bias in the included RCTs. The main MR analysis for maternal calcium also suggested no strong evidence of an effect on birth weight . Some sensitivity analyses suggested that the genetic instrument for calcium was associated with birth weight via exposures that are independent of calcium levels . Application of instrumental variable analyses to RCTs suggested that calcium has a substantial effect on birth weight  that was not consistent with any of the MR results. However, the RCT instrumental variable estimate may have been exaggerated because of risk of bias in the included RCTs. Other study limitations include the low response rate of UK Biobank, which may bias MR estimates, and the lack of suitable data to test whether the effects of genetic instruments on maternal calcium levels during pregnancy were the same as those outside of pregnancy.We performed two-sample mendelian randomisation (MR) using genetic instrumental variables associated with 25(OH)D and calcium that had been identified in genome-wide association studies D does not influence birth weight in otherwise healthy newborns. However, the effect of maternal circulating calcium on birth weight is unclear and requires further exploration with more research including RCT and/or MR analyses with more valid instruments. William Thompson and colleagues, using mendelian randomisation methods, reveal no association between levels of maternal circulating vitamin D and birth weight of newborns. Birth weight that is lower or higher than average has been associated with poor health outcomes across the life span, including infant mortality, cardiovascular disease, and type 2 diabetes.If we can identify modifiable maternal factors in pregnancy that are causally related to birth weight, we may be able to reduce the number of babies that are born with lower or higher than optimal birth weight. This may in turn help to reduce the associated poor health outcomes.Previous studies have suggested that higher maternal 25(OH)D and calcium in pregnancy are associated with higher birth weight. However, many of those studies used conventional multivariable regression in observational studies and may be subject to residual confounding. Few of them estimated the size of the effect of either maternal 25(OH)D or calcium levels on birth weight.We estimated the effects of maternal gestational 25(OH)D and calcium levels on offspring birth weight using mendelian randomisation, a method that uses genetic data to overcome certain limitations of traditional observational studies—for example, residual confounding.We analysed genetic data on 190,406 women from the UK Biobank who reported the birth weight of their first child, along with the results from published studies of genetic associations with 25(OH)D and calcium levels in 122,123 and 61,275 individuals, respectively.n = 4,576 mother–child pairs) and Exeter Family Study of Childhood Health , respectively.We checked that the results were not biased by offspring genotype using data from two UK birth cohorts, the Avon Longitudinal Study of Parents and Children  as an instrumental variable to estimate the effect of 25(OH)D or calcium on birth weight.We found no evidence of a strong effect of maternal 25(OH)D on birth weight.We found inconsistent evidence of effects of maternal calcium on birth weight.Our findings do not support using vitamin D supplementation during pregnancy or preconceptually to influence offspring birth weight.The effect of calcium on birth weight is still unclear and needs further investigation in well-powered genetic studies and/or well-conducted randomised controlled trials. Infants with lower or higher birth weight (BW) than average are at an increased risk of neonatal mortality and morbidity . BW is aMaternal gestational circulating 25(OH)D  and calcMendelian randomisation (MR) is a method in which genetic variants associated with a modifiable exposure are used as instrumental variables to estimate the causal effect of the exposure on an outcome . As geneThe analyses plan was developed by RMF, DAL, WDT, and MCB, prior to any analyses beginning. It was acted on by WDT and MCB. The analysis plan has not been published but was informally recorded in meeting notes. We made one change to the overall study plans after completing analyses; we undertook risk-of-bias assessment of the RCTs with a focus on factors that might mean instrumental variable assumptions were violated. This was motivated by differences in results comparing MR to instrumental variables in RCTs, particularly in relation to the effects of calcium on BW. We made one change to the plan following reviewers’ comments; we undertook two multivariable MR analyses to adjust for potential confounders. These were (1) multivariable MR of the association of 25(OH)D with BW adjusting  for maternal height and (2) partial multivariable MR of the association of calcium with BW adjusting  for maternal educational level.The study design and different data sources are summarised in We used two-sample summary data MR to explore the effect of maternal circulating vitamin D (25[OH]D) and calcium levels on offspring BW . We usedN = 273,495) were also asked to report the BW of their first child. Female participants that reported having a multiple first birth were excluded from our analyses . A total of 216,839 women with a singleton pregnancy for their first child reported the BW of their first child. Values were reported to the nearest whole pounds and were converted to kilograms by multiplying by 0.454 for our analyses. When women reported the BW of the first child at multiple time points , we used the mean of all measures after excluding any women with a difference of >1 kg between any two measures (N = 31). We further excluded from the whole sample any women who reported the BW of their first child < 2.2 kg or > 4.6 kg . This was done to reduce bias from reporting errors and, in relation to those < 2.2 kg, to exclude extreme preterm births, given that we do not have information on gestational age. BW of first child was regressed against the women’s reported age at first birth and UKB assessment centre location to reduce heterogeneity in reporting bias by these characteristics. Residuals from that regression model were then standardised to a mean of 0 and an SD of 1, with the standardised residuals being used in all analyses and final results converted back to grams. The analyses were done on standardised residuals to reduce the amount of computing power needed when doing the initial GWA analyses, and so the summary GWAS data that we used for our main analyses were already in this standardised format. UKB participants’ (women and men) reports of their own BWs were used in sensitivity analyses described below, and similar methods of exclusion and use of standardised (on age at assessment and centre) residuals, with conversion of results back to grams, were used for own BW  from obstetric clinical records at the time of birth, which occurred between 1991 and 1992 (mean [SD] age of mothers 29 [4.6] years) in ALSPAC  and betwThe GWAS of 25(OH)D and calcium that we have used in our MR analyses were undertaken on adult European-origin (nonpregnant) women and men. Our MR analyses assume that the magnitude of gene instrument variable-25(OH)D  association from those studies are the same in women during pregnancy. We were able to test this for 25(OH)D in ALSPAC, in which 25(OH)D was measured during pregnancy in mothers, using methods that have previously been reported  see S4 . NeitherFor UKB, we analysed data from the May 2017 release of imputed genetic data, which have been extensively described elsewhere . Given tp-value > 1 × 10−6), passed QC in all 106 batches, and were present on both genotyping arrays were included in the genetic relationship matrix (GRM). For the GWA analyses of BW of the first child , the genotyping array and genotyping release (interim versus full) were included as covariates in the regression model. For the GWAS of participants’ own BW , genotyping array, age at baseline, and sex were adjusted for in all models.To account for population structure and relatedness, a linear mixed model implemented in BOLT-LMM v2.3  was used−6, sex mismatch, kinship errors, and 4.56 SD from the cluster mean of any subpopulations cluster) D supplement or placebo/other control), together with their respective CI values, from 24 RCTs that presented results for both of these (g pairs)  (S3 Tablg pairs)  (S4 Tablg pairs)  provided2 statistic and leave-one-out analysis. For the vitamin D supplementation, RCTs’ differences in mean 25(OH)D were in nmol/l, and for the calcium supplementation, RCTs’ differences in mean calcium were in mg/dl. In order to make the instrumental variable analyses in RCT results comparable with our two-sample MR results, for 25(OH)D we assumed that a 10% difference was equivalent to 6.2 nmol/l , and we multiplied the Wald ratio estimates by 6.2 to scale the results so that they represented the difference in mean BW per 10% increase in 25(OH)D. We did the same with calcium but used 0.5  to scale results so that they represented the difference in mean BW per 1 SD of calcium.We calculated the Wald ratio estimate for each RCT by dividing the difference in mean BW by the difference in mean 25(OH)D  and then pooled these using a fixed-effect meta-analysis and tested for between-study heterogeneity using the IThe characteristics of included participants from UKB, ALSPAC, and EFSOCH are shown in 2 = 0.0%), and results were consistent in leave-one-out analysis with the main results  and with each other (p = 0.981) per 10% increase in maternal circulating 25(OH)D. However, the RCT instrumental variable effect was 5.94 g  per 10% increase in maternal circulating 25(OH)D (2 = 16.2%), and results were consistent with the main overall result and with each other in leave-one-out analysis D on offspring BW, with consistent null findings in the main Wald ratio and IVW analyses and also with the MR-Egger and weighted-median results using the UKB data . There wch other . MR effech other . The effch other . Our mai 25(OH)D . There wanalysis .The difference between the MR and instrumental variable analyses in RCTs is small, and the CIs for the two results overlap, suggesting that they are statistically consistent. To explore this small difference further, we undertook post hoc risk-of-bias assessment of the 24 RCTs. Though there were only a small number of genetic instruments, and thus limited ability to test potential biases from horizontal pleiotropy and weak instruments, there was no strong evidence that 25(OH)D synthesis or metabolism had an effect on BW .p = 0.025) and the effect estimate suggesting a possible modest effect of 41 g  per 1 SD increase in maternal circulating calcium, compared to the weak inverse effect in the IVW and Wald ratio estimates of −20 g  (2 = 62.4%), and removing one of the SNPs (rs1801725) strengthened the inverse effect estimate to −61 g  per 1 SD increase in maternal circulating calcium, with removal of the other six SNPs resulting in estimates close to the main result and consistent with each other (p = 0.558) per 1 SD increase in maternal circulating calcium. There was no evidence that fetal genotype influenced the results, as results were similar with or without adjustment for it, and the IVW analyses for adults’ own BW gave a similar result to the IVW analyses for first child’s BW (p = 1.43 × 10−9) change in BW per 1 SD increase in maternal circulating calcium (2 = 0.0%), but the main result and leave-one-out analyses had wide CIs (Ns = 23 to 72), and the two largest RCTs (N = 120 and 274), together with two other studies, did not use intention-to-treat analysis (N = 120) was also judged by the authors of the systematic review to have high risk of bias (or it was noted that there was insufficient information to determine risk) across all of the specific domains they assessed  . There wch other . The Walild’s BW . The RCT calcium . There wwide CIs . Given tanalysis , which cassessed .The magnitude of the 25(OH)D SNPs association with maternal circulating 25(OH)D in ALSPAC was similar to the magnitudes reported in the source GWAS for five of the seven SNPs. There were differences for rs8018720 (which was weakly inversely rather than positively associated with 25[OH]D in ALSPAC) and for rs117913124 (a weaker positive difference in ALSPAC when compared to source GWAS) . DespiteNeither the 25(OH)D nor the calcium genetic instrumental variable–weighted allele scores were associated with observed confounders in the UKB, ALSPAC, or EFSOCH, with the exceptions of maternal height, education, and Townsend deprivation index (TDI). The score for 25(OH)D was negatively associated with height in the UKB, positively associated with height, and negatively associated with TDI in EFSOCH and showed no association in ALSPAC; the score for calcium was negatively associated with educational level and positively associated with TDI in UKB . Given tThis study triangulated two approaches  assessing the effects of maternal circulating 25(OH)D and calcium on BW. Across the main and all sensitivity MR analyses, we found no evidence that maternal 25(OH)D has an important effect on BW, but applying instrumental variable analyses to RCTs, there was evidence of a weak positive effect. Findings for maternal circulating calcium were inconsistent across methods and sensitivity analyses, making it difficult for us to conclude from the data used here what the effect of maternal circulating calcium on BW is.p = 0.002) higher BW per 10% increase in 25(OH)D, this might be exaggerated by limitations of the original RCTs  [Although the instrumental variable applied to RCT analyses for 25(OH)Ds effect suggested a weak positive effect of 5.94 g  . When cof birth) , which if birth) . That saf birth) . As preef birth) , these fIt is possible that low circulating 25(OH)D levels are not a suitable indicator of vitamin D deficiency, as 1,25(OH)2D is the biologically active form of vitamin D and remains within reference limits even when circulating 25(OH)D levels are low, suggesting valid genetic and RCT instruments associated with 1,25(OH)2D levels could give different results. However, we are not aware of such instruments, and circulating 25(OH)D has a longer half-life in the body, making it a more stable measurement and possibly a better indicator of long-term vitamin D exposure . FurtherCASR) gene, which codes for the calcium-sensing receptor protein and is widely expressed in the parathyroid gland, kidneys, and intestines and regulated blood levels of calcium [CASR rs1801725 alone and the MR-Egger result, all suggest that our main IVW results might be biased by masking horizontal pleiotropy and that there is a modest positive causal effect of calcium on BW. Possible sources of masking pleiotropy could be alterations to glucose levels, as one of the genetic instruments (rs780094) is associated with fasting glucose, which is known to influence BW [Although the main MR analyses did not support a causal effect of maternal calcium on offspring BW, sensitivity (MR-Egger and leave-one-out) analyses were not completely consistent with the main MR effect estimates, and the instrumental variable analyses applied to RCTs were markedly different from any of the MR results. There are a number of possible reasons why we might have found these differences. The MR estimate may be biased by masking horizontal pleiotropy, which both our MR-Egger and leave-one-out analyses suggest. Thus, the MR-Egger effect estimate suggests a modest positive effect with wide CIs, and the leave-one-out analyses suggest that the heterogeneity between individual SNP Wald ratios is driven by one SNP, rs1801725, which when removed results in an inverse association (again suggesting masking pleiotropy). rs1801725 is in the calcium-sensing receptor . On the other hand, for in utero exposures, timing rather than duration of exposure might be important . Though 2 for one of these seven SNPs (rs1801725) [2 in our study suggesting the genetic variants explained approximately 3% of variation in circulating 25(OH)D compared with <1% of the variation explained by the two SNPs used in the previous MR study [To the best of our knowledge, this is the first study to compare results from MR and instrumental variable analyses applied to RCTs to investigate the effect of maternal circulating calcium levels on offspring BW. The GWAS that we used for genetic instrument–exposure associations (sample 1) in our two-sample MR did not provide the percentage variation in calcium that all seven SNPs explained. However, based on a previous study that provided the R1801725) , we know1801725) . These aMR study . We haveKey potential limitations include the low response to UKB and maternal report of first child’s BW many years after their birth. Recent research suggests that a highly select cohort (as in the case of UKB with a 5% response ) can resIn UKB, the 25(OH)D genetic risk score was negatively associated with height, and as it is possible that greater maternal height results in greater BW, independently of the offspring genotype , this coHorizontal pleiotropy is a key source of bias for MR studies. For the effects of maternal circulating 25(OH)D on BW, the consistency of results across all of our MR analyses suggests that this has not been a key source of bias for that exposure. We did find evidence of potential horizontal pleiotropy with maternal calcium effects and as discussed above and acknowledge that further studies are required to explore that effect. We have assumed that genetic variants that relate to 25(OH)D/calcium in (nonpregnant) women and men have the same magnitudes of association with these exposures in pregnant women. We were unable to test this for calcium but did show that five out of the seven variants related similarly to circulating 25(OH)D during pregnancy as in the GWAS. Removal of either of the two that did not associate as strongly to 25(OH)D in pregnancy in our leave-one-out analyses was consistent with the main results and all other leave-one-out analyses. Although we aimed to use randomisation to use supplementation with vitamin D or calcium as an instrumental variable for RCTs, some studies did not provide results from intention-to-treat analyses. Therefore, results from these studies could be biased by noncompliance. The aim of triangulation is to compare results from different methods that have different key sources of bias. Although both approaches that we have used here rely on instrumental variables and have the same underlying assumptions, the sources of violation of these assumptions differ between MR and instrumental variables applied to RCTs. For MR, horizontal pleiotropy is the key source of bias, whereas for instrumental variables applied to RCT, it is lack of concealed random allocation (which can introduce bias) and not using intention-to-treat analysis (which can produce a path from the instrument to outcome and/or introduce confounding). The fact that there was no marked difference between our MR and RCT analyses strengthens confidence in the findings for 25(OH)D. However, as discussed above, the different results for calcium need further exploration. In both our two-sample MR and instrumental variable analyses applied to RCTs, we have used summary data and are unable to explore possible nonlinear effects, though observational studies do not suggest that these are present. Similarly, we cannot explore effect modification, but we are not aware of any evidence that the effects that we have looked at do differ by other characteristics. Our MR analyses are in white European populations and may not generalise to other groups, such as those with different levels of skin pigmentation and exposure to sunshine. However, most of the RCTs for vitamin D and calcium supplementation are in non-Europeans. In particular, most vitamin D RCTs were in Middle Eastern or South Asian populations, and a recent RCT in Bangladesh, which was not included in the systematic review that we used in this paper for the RCT analyses, found that vitamin D supplementation during pregnancy made no difference to BW . Thus, eIn conclusion, our results suggest that maternal circulating 25(OH)D does not have a clinically important effect on BW, so pregnancy supplementation with vitamin D is unlikely to affect mean BW. Higher maternal circulating calcium may increase BW, but further research is required to clarify this, including larger samples for undertaking pleiotropy ‘adjusted’ MR analyses and larger, better-conducted RCTs of calcium supplementation.S1 Checklist(DOCX)Click here for additional data file.S1 Text(DOCX)Click here for additional data file.S2 Text(DOCX)Click here for additional data file.S3 Text(DOCX)Click here for additional data file.S4 TextALSPAC, Avon Longitudinal Study of Parents and Children.(DOCX)Click here for additional data file.S5 Text(DOCX)Click here for additional data file.S1 Table(PDF)Click here for additional data file.S2 TableSNP, single-nucleotide polymorphism.(PDF)Click here for additional data file.S3 TableRCT, randomised controlled trial.(PDF)Click here for additional data file.S4 TableRCT, randomised controlled trial.(PDF)Click here for additional data file.S5 TableSNP, single-nucleotide polymorphism.(PDF)Click here for additional data file.S6 TableSNP, single-nucleotide polymorphism.(PDF)Click here for additional data file.S7 TableALSPAC, Avon Longitudinal Study of Parents and Children; EFSOCH, Exeter Family Study of Childhood Health; SNP, single-nucleotide polymorphism.(PDF)Click here for additional data file.S8 TableIV, instrumental variable; RCT, randomised controlled trial.(PDF)Click here for additional data file.S9 TableIV, instrumental variable; RCT, randomised controlled trial.(PDF)Click here for additional data file.S10 Table(PDF)Click here for additional data file.S11 TableMR, mendelian randomisation.(PDF)Click here for additional data file.S1 FigALSPAC, Avon Longitudinal Study of Parents and Children; EFSOCH, Exeter Family Study of Childhood Health.(PDF)Click here for additional data file.S2 Fig(PDF)Click here for additional data file.S3 FigRCT, randomised controlled trial.(PDF)Click here for additional data file.S4 Fig(PDF)Click here for additional data file.S5 FigRCT, randomised controlled trial.(PDF)Click here for additional data file.S6 Fig(PDF)Click here for additional data file.S7 Fig(PDF)Click here for additional data file.S8 FigRCT, randomised controlled trial.(PDF)Click here for additional data file.


We thank Panesar and Fernandez-Miranda  for theisystem,” as the central part of its trajectory  is in close relationship with the stems of the three superior longitudinal fasciculus. And we fully agree that the terminology “superior longitudinal fasciculus” should not be used for the arcuate fasciculus, as the whole shape of the tract is not longitudinal.With this general principle in mind, it seems clear that the arcuate fasciculus belong to the broad category of “superior longitudinal Regarding the comment about the cingulum, we do not really understand the criticism, as we fully concur with their statement. We were among the first to deny the existence of the SLF I as initially described by Makris  temporo-parietal fasciculus is separated from the vertical occipital fasciculus. But again, the location and orientation of their stems makes intuitive their assignment to a common category that we called “posterior transverse system,” thus mirroring the anterior transverse system.As imperfect and incomplete as it is, our taxonomic nomenclature has the merit to be a recursive, anatomically-based, and flexible approach to the hard task of classifying and designing the complex sets of connections of the human brain and we hope it could help naïve people to the field to assimilate more easily our current state of knowledge.EM wrote the response to the commentary. SS and LP approved it for publication.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.