Macro- and micronutrient deficiencies as well as exposure to tobacco, alcohol, and certain medications during gestation have been shown to negatively impact nephrogenesis and reduce one’s nephron number. In 1988, Brenner et al. put forth the controversial hypothesis that a reduced nephron complement is a risk factor for hypertension and chronic kidney disease in adulthood. Since then numerous animal and human studies have confirmed this relationship demonstrating that there is an inverse association between blood pressure and nephron number. As our understanding of the developmental programming of hypertension and other non-communicable diseases improves, more effective preventive health measures can be developed in the future.Primary hypertension continues to be one of the main risk factors for cardiovascular disease worldwide. A stable intrauterine environment is critical for the future development and health of the fetus. The developing kidney has been found to be especially vulnerable during this time period, and epidemiological studies have demonstrated that an adverse  Primary hypertension is one of the leading risk factors for morbidity and mortality in the world, and it has been designated as the primary risk factor for the global disease burden , 2. ApprRecently, attention has been paid to the contribution of the intrauterine environment to the development of chronic and non-communicable diseases. Epidemiological studies have demonstrated that a poor intrauterine environment is associated with an increased risk of hypertension, chronic kidney disease, and diabetes –9. The dvia the secretion of growth factors, resulting in the development of the collecting duct system , 2220, 2ffspring . The folBirth weight has been identified as the primary determinant of nephron number . Low birIn addition, racial and ethnic populations with a high prevalence of low birth weight have been found to have fewer nephrons or smaller kidneys –27. Babifgf7 and fgf10), glial cell-derived neurotrophic factor , hepatocyte growth factor , transforming growth factor β , and signal transduction proteins from the Wnt family , that is associated with an increased total kidney volume in newborns when adjusting for body surface area. The ALDH1A gene is involved in the metabolism of retinoic acid, and infants who were homozygous for this variant were found to have higher umbilical cord blood levels of retinoic acid .Maternal nutrition plays an essential role in the development of the fetus. The developing kidney is especially vulnerable to the effects of a poor maternal diet, and maternal malnutrition during pregnancy has often been proven to be associated with suboptimal renal phenotypes among her offspring . Macro- Protein restriction during gestation significantly reduces nephron number –41. WoodMicronutrients also significantly impact renal organogenesis , 22. GilThe detrimental effects of chronic alcohol exposure to the fetus during pregnancy have been well documented. However, the effects of acute prenatal alcohol exposure are less understood, especially its influence on the developing kidney. Gray et al. investigated the effects of acute prenatal ethanol exposure on nephron endowment in rats. Either ethanol or saline was administered to Sprague-Dawley rats on embryonic days 13.5 (E13.5) and 14.5 (E14.5). Ethanol exposed rodents were found to have a 10–20% reduction in nephron number as well as a higher mean arterial blood pressure. In addition, E15.5 rats were found to have a reduced gene expression of significant regulators of ureteric branching morphogenesis . Fewer ureteric branch points and tips were also found in embryonic kidneys that were cultured in media containing ethanol .Tobacco exposure is another established teratogen, which has been demonstrated to be negatively associated with birth weight and positively associated with blood pressure , 46. Taain vitro and in vivo (Elevated levels of maternal corticosteroids during pregnancy have been shown to be associated with a low nephron complement and higher blood pressure among her offspring . Interes in vivo . Ureteri in vivo .Exposure to non-steroidal anti-inflammatory drugs has been shown to have variable effects on one’s nephron number . RodentsOne of the initial proponents of the fetal origins of disease was Dr. David Barker. In the 1980s, Barker and colleagues discovered an inverse relationship between birth weight and coronary artery disease mortality rates in a cohort of men and women born in Hertfordshire, United Kingdom. They observed a twofold increase in mortality rates when comparing individuals with the lowest birth weight with those with the highest birth weight , 51. Barin utero environment (The theory of “developmental programming” or the “developmental origins of health and disease,” states that environmental influences during sensitive periods of development can result in permanent alterations of function and structure of an organism . In ordeironment . As descironment , 44, 48.In 1988, Brenner et al. put forth the hypothesis that a low nephron number is a risk factor for adult hypertension Figure . They prSince initially hypothesized over 30 years ago, there have been a number of animal and human studies that have sought to determine the relationship between nephron endowment and hypertension.Through animal studies, it is becoming increasingly clear that a relationship exists between nephron number and hypertension. A mouse model of prematurity found that those delivered 1 and 2 days early had 17.4 and 23.6% fewer nephrons, respectively, compared to full-term mice, and the premature mice subsequently developed hypertension . In addiHuman epidemiological studies have used birth weight as an indicator of fetal nutrition status and therefore a surrogate for lower nephron endowment. Systematic reviews and meta-analyses of the literature have consistently shown an inverse relationship between birth weight and blood pressure. Law et al. reviewed 34 cohort studies published before 1996. After adjustment for body mass index, regression coefficients of birth weight indicated a negative association with systolic blood pressure in 26 studies, indicating a 2–3 mmHg/kg reduction in systolic blood pressure in children and 2–4 mmHg/kg reduction in adults with increasing birth weight . In a reThe relationship between birth weight and blood pressure in childhood is more complicated, and studies published since 2000 have shown inconsistent results muddied by poorly powered studies focusing on specific populations. There is also some evidence that the inverse relationship between blood pressure and birth weight becomes more pronounced as the age of the study population increases, although this finding is weak and may not be statistically significant . StudiesThere is a significantly higher prevalence of hypertension in African-Americans, which is likely multifactorial in etiology. African-Americans are also at higher risk for being low birth weight. This suggests that being low birth weight could be one of the factors implicated in predisposing African-Americans to higher rates in hypertension later in life .Since most of the studies in this area focused on Caucasians, there is little published on the African-American population. Hulman et al. evaluated 137 urban African-American adults and found no relationship between birth weight and adult blood pressure . Donker Interestingly, hypertension is also highly prevalent among Australian Aboriginals and it appears that a reduced nephron complement mediates the relationship between low birth weight and hypertension within this population . Hoy et Low birth weight has been shown to impact the nephron number at birth; however, numerous factors affect nephron number throughout one’s lifetime. Environmental factors such as weight gain and aging later change nephron number and therefore risk , 77. HowNephrons are hypertrophied atrophy sooner, therefore decreasing the nephron number even more so . Though There is an increasing body of literature demonstrating the relationship between nephron number and hypertension. These studies show that there is an inverse relationship between nephron endowment and hypertension. As this relationship is established, factors that contribute to determining nephron number are being elucidated. Genetics clearly plays a role in nephron endowment, one that possibly explains racial differences in nephron number at birth. Environmental factors such as birth weight, body size, and age also have an impact. Having a way to assess for life-long risk of renal and CVD could be instrumental in initiating preventive health measures earlier in life. Like genetics, low birth weight is a predetermined risk factor that could be utilized for risk stratification and thus early initiation of preventive health measures. Recent advances in accurate and precise estimates of glomerular number will improve our understanding of the factors contributing to nephron endowment and its implications for future health.SG, AT, and CS contributed equally to the writing of this manuscript.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.


This paper examines the effect of maternal socio-economic status in the household, such as their autonomy, ability, freedom and bargaining power, on child birth registration in India using the nationally representative India Human Development Survey-II (IHDS-II), 2011–12. We have estimated a multilevel mixed effects model which accounts for the hierarchical structure of the data and allows us to examine the effects of unobserved ‘district’ and ‘community’ characteristics along with individual child level characteristics on probability of birth registration. The results show that between-districts and between individuals differences share a considerably high and an almost equal proportion of the variations in probability of birth registration in India. At individual child level, mother’s socio-economic status such as her ability to move around independently and her exposure to outside world, significantly raise the probability of birth registration. More importantly, the marginal effects of the maternal autonomy indicators: mother’s ability to move around freely and her control over resources, on birth registration vary across districts in India. Other variables such as institutional birth, mother’s antenatal care seeking behaviour, caste, religion, household wealth and parental education are significant determinants of birth registration. We, therefore, estimate a comprehensive model which controls for household and individual child level covariates at level 1 including mother’s social and economic status, state and community level contextual characteristics. We also allow for two random slopes at the district level for mother’s social and economic status indicators–namely, mother’s autonomy and mother’s ability. A random slope at the community level is also included for ‘proportion of institutional delivery at the village level’. The results from estimation of this model are shown in Our primary focus of analysis in this paper is the variables representing mother’s social and economic status and bargaining power in the household. The summary variable  representing ‘mother’s ability’ is significantly associated with birth registration. Mothers that are able to visit health centres, friends/relatives, a local shop or travel short distance by train/bus on their own are more likely to have their children registered. This variable reflects the mother’s capacity to leave the house unaccompanied and move around without needing a chaperon. Higher maternal mobility is related to greater decision-making ability within the household . As primWe also find that two other variables which represent mother’s exposure to outside world–namely, whether the mother had been to a metro/another state/abroad in the last 5 years; and whether she had gone out on family outings- are significant determinants of birth registration. In particular, having a mother who had been to a metro/another state/abroad increases the odds of birth registration by a factor of 1.18. A mother travelling to another state or abroad signifies [that she enjoys] more autonomy in the household . And undAdditionally, we had included a random slope coefficient for ‘mother’s autonomy’ at the district level. The ‘mother’s autonomy’ variable shows mother’s decision-making power on issues such as purchase of property/land, wedding expenses, what to do if a child falls sick and whom the child should marry. A mother that enjoys a greater degree of ‘autonomy’ would have greater access and control over economic resources in the household. It is documented in the literature that autonomous mothers would allocate more resources towards their children , 42. ThuAmong the child specific factors included in the regression, place of birth of child is statistically significant. The odds of birth registration increase by a factor of 4.6 for a child who has had an institutional birth. We can also see that institutional delivery has one of the largest impacts on the likelihood of birth registration in our model. The high likelihood of registration for institutionally delivered children can be attributed to the fact that the reporting of such births to the ‘Registrar of Births and Deaths’ is the responsibility of the medical officer in charge. More importantly, medical officers who attend the birth of a child are obligated to report the incidence to the Registrar . Brito eMother’s number of antenatal visits at last pregnancy is another significant variable positively associated with birth registration. The odds of birth registration for mothers who have had four or more antenatal check-ups is 2.18 times that of mothers who have had no check-up. Antenatal care provides pregnant women with education, counselling, screening and treatment to ensure mother and foetus remain in good health . MoreoveOur results also indicate that household wealth is significantly associated with birth registration. The odds of birth registration for households in the top wealth quintile is 1.49 times that of households in the bottom quintile. Other cross-country studies have also established the positive impact of wealth status on birth registration , 47. ThiMoreover, the dummy variable for religious background shows that the odds of birth registration for a child born in a Muslim family is 0.80 times that of a child born in a Hindu family. The public health literature in India which has examined the Hindu-Muslim differences in fertility planning may shed some light into this result. There are studies that indicate higher level of unmet need for family planning among Muslims . Singh eWe also find that a child belonging to Scheduled Castes (SC), Scheduled Tribes (ST) or other castes has a lower probability of registration compared to the Forward/General castes. Since India’s independence in 1947 the social caste structure has been identified as a significant hindrance for the socioeconomic development of minority groups. Despite government effort to empower lower caste groups and minorities, significant disparities across all indicators of development continue to exist. Studies indicate that social caste structure in India is a significant predictor for safe delivery and post-natal care utilization with women from Scheduled Castes (SC) and Scheduled Tribes (ST) less likely to access such services . Thus, iAmong the community level factors included in our model, mean years of schooling and proportion of institutional delivery come out as significant. An increase in mean years of schooling by one year is associated with 1.1-fold increase in the odds of birth registration. Thus, education is important not only at the household level but also at the community level. Living in a community with a higher proportion of institutional births also raises the probability of birth registration. These results point out that there are advantages that spill over to the individual/household from living in a community with high mean years of schooling and institutional delivery. We have included a random slope coefficient for proportion of institutional delivery at community level to allow for the effect of this variable (on birth registration) to vary across communities in India but this didn’t come out as significant.Living in non-low income states has the largest positive effect on birth registration in our regression. The odds of birth registration for a child living in a non-low income state is 9.2 times that of a child living in low income state. This might be reflecting the poor state of health and other infrastructure leading to poor service delivery in low income states in India. Chotia and Rao (2015)  note thaThis paper examines the determinants of birth registration in India using a multilevel hierarchical mixed model. In particular, we looked at mother’s autonomy, ability and bargaining power within the household and its significance for child birth registration. The rationale behind using a multilevel mixed effects model is the hierarchical nature of the data in that child and households are nested within communities, which in turn are nested within districts. The estimation results provide us with useful policy implications to increase birth registration in India. Our results show that 45 percent of the variation in birth registration in India lies between individual level differences and 44 percent lies between district level differences and remaining 11 percent lies between community level differences. This brings the policy focus on individuals and districts for targeting birth registration. At individual level the results indicate that the summary variable representing mother’s ability increases the probability of birth registration. Ability to move around independently is an important trait for the mothers because such mothers do not have to wait for their husbands to take their children for immunisation, health check-ups, birth registration…etc. Two variables representing mother’s exposure to outside world  are also significant. So, mothers that have better bargaining power in the household and are exposed to liberating and progressive ideas from their surrounding that they could use to advance children’s welfare are more likely to register the birth of their children. Again, random slopes at the district level for two indicators of mother’s bargaining power- namely, mother’s ability and mother’s autonomy- came out as significant confirming that while these variables significantly influence birth registration, the level of their influence varies across districts in India. So, policies targeting to empower mothers to improve maternal and child health outcomes and birth registration should have a district level focus.Our estimates further showed that institutional delivery is a highly significant determinant which increases the probability of birth registration. This is consistent with a priori expectation as medical officers presiding over delivery are duty bound to report the birth. The number of mother’s antenatal visits also came out as significant, which is not surprising given that mothers will also get advice on postpartum care when they access antenatal services. Finlayson and Downe (2013)  note thaRobertson (2013)  notes thOur study also established that probability of birth registration increases by wealth quintile. The direct and indirect costs associated with the process of registration may discourage households in lower wealth quintiles from having children registered. We also find disadvantaged groups such as Scheduled Castes (SC), Scheduled Tribes (ST) or other castes have a lower probability of birth registration compared to the Brahmin and Forward/General castes. Children in low income states are also less likely to be registered. This suggests that policy should give special attention to raising birth registration rates of these disadvantaged groups and states.


Dentigerous cyst is the most common odontogenic cyst. It is characterized by a unilocular radiolucent lesion that encloses permanent tooth buds or, under certain circumstances, displaced tooth buds. Buccal bony expansion and a missing tooth is the most common clinical feature. Various treatment modalities have been mentioned in the literature for management of dentigerous cysts. This article presents a left mandibular dentigerous cyst in a 10-year-old boy. Marsupialization was the treatment of choice and a denturelike space maintainer was provided. Long-term follow-up revealed good healing of the bony lesion.Bhardwaj B, Sharma S, Chitlangia P, Agarwal P, Bhamboo A, Rastogi K. Mandibular Dentigerous Cyst in a 10-Year-Old Child. Int J Clin Pediatr Dent 2016;9(3): 281-284. The clinical findings are cortical bone expansion, adjacent permanent tooth bud displacement, and root dilacerations. The developmental type is usually found in the late 2nd and 3rd decades. It occurs in mature teeth, generally without inflammation. The inflammatory type is found in the 1st and early 2nd decades. It usually occurs in a nonvital immature deciduous tooth.7Dentigerous cyst is the most common odontogenic cyst. It is associated with the crown of an unerupted tooth. The cyst cavity is lined with reduced enamel epithelium derived from the tooth-forming organ.A 10-year-old boy was referred to the department of oral and maxillofacial surgery for evaluation of pain and swelling in the left mandibular vestibule. Physical examination revealed a hard submucosal mass with buccal cortical plate expansion in region of second premolar. A panoramic radiograph showed a well-defined radiolucent lesion, measuring approximately 3 × 2 cm in diameter and including the crowns of the unerupted mandibular left 2nd premolar. The root of the adjacent canine, 1st premolar, and 1st molar were also involved in the lesion. The root apices of these teeth were still not closed. No signs of root resorption were present . ConsideThe patient was treated with extraction of the 2nd premolar, marsupialization of the cystic lesion  by creat8 It can cause cortical plate expansion and result in facial asymmetry.9 A unilocular well-defined radio-lucency lesion and asymptomatic lesion characterized on radiographic examinations.2 Other radiolucent lesion, such as radicular cyst, odontogenic keratocyst, ameloblastoma, and odontoma should be differentially diagnosed with dentigerous cyst.10Dentigerous cyst is associated with crown of unerupted teeth.211The pathogenesis of dentigerous cyst is by the accumulation of fluid either between the reduced enamel epithelium and the enamel or between the layers of the enamel organ. This fluid accumulation might be a result of pressure exerted by a potentially erupting tooth on the follicle, which obstructs the venous outflow and induces serum transudation across the capillary wall.12 Shaw et al,13 Benn and Altini,2 Bando and Nagayama,14 and Aguiló and Gandía.15In 1928, Bloch-Jorgensen, in a series of 22 cases of dentigerous cysts, reported a diseased deciduous tooth or a remnant found in direct contact with the cyst wall. This finding has also been pointed out by Azaz and Shteyer,16 has demonstrated that dentigerous cyst is caused by an alteration of the normal development of the tooth germ related to mechanical obstruction or an eruptive deviation.16Sharp and Helsper20 If preservation of the displaced teeth is desirable, marsupialization is a rather conservative treatment option.26 Marsupialization is the conversion of a cyst into a pouch by suturing the cyst lining to the oral mucosa. This method has fewer complications than enucleation regarding the preservation of important anatomical structures and developing permanent tooth germs.Various treatment modalities include complete enu-cleation and marsupialization. The choice of treatment depends on various factors, such as age of the patient, location of the cyst, tooth position in relation to the cyst, and the degree of the axial inclination and root formation. If the cyst is associated with a supernumerary tooth, complete enucleation of the cyst along with extraction of the tooth may be the first choice.21634 which may consist of enucleation and extraction of the teeth embedded in it or affected by it. In very large cysts, an initial phase of marsupialization of the lesion to the oral cavity followed by enucleation is also recommended. Although Kaban34 only mentions occasional tooth-saving as a possibility, he does state that thorough enucleation should never be compromised.The treatment of dentigerous cyst is usually surgi-cal,38However, there are various single case reports where the cyst was opened to the oral cavity and a stent - either a rubber tube, a removable device, or a gauze packing -was used to keep the opening patent.39 These factors are significant in cases of large dentigerous cyst in children and presents a better prognosis for the teeth involved in the lesion.Children have greater capacity to regenerate the bony structure compared to adults; moreover, teeth with open apices have a great eruptive potential.The choice of treatment for dentigerous cyst is ruled by various factors, such as age of patient, location and size of the cyst, tooth position in relation to the cyst, proximity to the vital structures, and degree of the axial inclination of the tooth and its root formation. Marsupialization stands as a favorable or preferred treatment modality in young patients.


Yet, CA is rarely used in epidemiologic studies in low- and middle-income countries (LMICs), which may bias population estimates of childhood undernutrition. To evaluate the effect of accounting for GA in the application of growth standards, we used GA-specific standards at birth (INTERGROWTH-21st newborn size standards) in conjunction with CA for preterm-born children in the application of World Health Organization Child Growth Standards postnatally (referred to as ‘CA’ strategy) versus postnatal age for all children, to estimate mean length-for-age (LAZ) and weight-for-age (WAZ) z scores.At birth (n = 4066), mean LAZ was higher and the prevalence of stunting (LAZ < −2) was lower using CA versus postnatal age (mean ± SD): − 0.36 ± 1.19 versus − 0.67 ± 1.32; and 8.3 versus 11.6%, respectively. Odds ratio (OR) and population attributable risk (PAR) of stunting due to preterm birth were attenuated and changed inferences using CA versus postnatal age at birth ; differences in inferences persisted at 3-months. At 12, 24, and 48-months, preterm birth was associated with stunting, but ORs/PARs remained attenuated using CA compared to postnatal age. Findings were similar for weight-for-age Population-based epidemiologic studies in LMICs in which GA is unused or unavailable may overestimate the prevalence of early childhood undernutrition and inflate the fraction of undernutrition attributable to preterm birth.The online version of this article (10.1186/s12982-018-0070-1) contains supplementary material, which is available to authorized users. Early childhood growth is an important indicator of a child’s future health and developmental potential , 2. ChilThe World Health Organization Child Growth Standards (WHO-GS) are the most widely accepted international normative standards for evaluating postnatal growth among children less than 5 years of age , 17. TheThe importance of accounting for GA in the timescale for evaluating neonatal outcomes is well-recognized in perinatal epidemiology , and theTo quantify the implications of not accounting for GA at birth in the application of child growth standards for population-based estimates of undernutrition in LMICs, we compared the use of a ‘CA’ strategy, which applied GA-specific standards at birth in conjunction with WHO-GS using CA for preterm-born children postnatally, versus postnatal age in the application of WHO-GS for all infants , to estimate mean nutritional indices and indicators  in a population-based cohort of term- and preterm-born children less than 5 years of age in Brazil. Additionally, we aimed to estimate the proportion of stunting and underweight attributable to preterm birth in the first 5 years of life using CA versus postnatal age strategy.Anthropometric data from the 2004 Pelotas Birth Cohort study were used. The cohort study methods have been described in detail elsewhere , 28. BriGA at birth was measured by three different methods: (1) the Dubowitz score based on the physical and neurological characteristics of the newborn; (2) date of the last menstrual period (LMP), as reported on a mother’s prenatal card or self-reported during perinatal interview; and (3) ultrasonography evaluation performed before 20 weeks of pregnancy, as recorded on the mother’s prenatal card . Althougn = 215) if birthweight-for-GA was outside the range of (1) ± 2 standard deviations (SD) according to the Fenton growth curves  . For exaMethods used for anthropometric assessment of children in the Pelotas 2004 cohort were previously described . Brieflyz scores were derived using 0 day when applying the WHO-GS and CA when applying the IG-NS  to permit application of the IG-NS and minimize the loss of data at birth. For one female infant born at 231 days GA, we used 232 days GA to derive z scores at birth because IG-NS do not estimate z scores for female infants born specifically at 231 days GA.At birth, length-for-age (LAZ) and weight-for-age (WAZ) -NS Fig. . Since tweeks GA , we usedweeks GA , and truIn the postnatal period beyond day 0, LAZ and WAZ were derived using the WHO-GS based on either the (1) postnatal age for all children, or (2) postnatal age for term-born children and CA for preterm-born children. In primary analyses, we did not use CA at postnatal ages for children born at early term GA  or late term GA  because the WHO-GS are based on a reference population of children born between 37 and 42 weeks GA and are therefore intended to be applied directly to all children born in the 5-weeks span of ‘term’ gestation. However, as a sensitivity analysis we used CA for all children (including term-born children) in the application of WHO-GS at postnatal follow-up visits.www.who.int/childgrowth/software/en/) [https://intergrowth21.tghn.org/global-perinatal-package/intergrowth-21st-comparison-application/) were used for the application of the reference/standard at birth.In additional sensitivity analyses, LAZ and WAZ for preterm-born children at the 3-month visit  were derived using the INTERGROWTH-21st postnatal standards for preterm-born children based on the Preterm Postnatal Follow-up Study (IG-PPFS) . The WHOare/en/)  and the n = 56), twins (n = 80), or newborns with missing or implausible GA (n = 42) or missing both weight and length at birth (n = 2) were excluded from this study , outside ± 2 months of the 12- and 24-month visits (n = 20), and outside ± 6 months of the 48-month visit (n = 85) were excluded. Infants born at < 240/7 weeks GA (n = 2) were excluded from analyses at birth as they were below the range of application of the IG-VPBR.Stillbirths (z score (WLZ) < −2SD) at each visit. Paired Student’s t test and McNemar’s test for paired proportions were used to evaluate differences in means and proportions, respectively. The odds of undernutrition among preterm-born children (exposed) relative to term-born children (unexposed) at each follow-up visit were estimated using unadjusted logistic regression. We also determined the population attributable risk percent (% PAR) of undernutrition due to preterm birth using the following formula [Repeated cross-sectional analyses were conducted to estimate and compare mean LAZ and WAZ derived using CA versus postnatal age in the application of size/growth standards at birth, 3, 12, 24 and 48-month visits. Indicators of undernutrition were the proportion of children classified as stunted (LAZ < −2 SD), underweight (WAZ < −2 SD), and wasted .n = 482; 11.7%) had a mean ± SD GA of 34.4 ± 2.5 weeks (median (interquartile range): 35 (2.0) weeks). Of the 482 children born preterm, 19.2% (n = 92) were born at ≤ 336/7 weeks and 81.3% (n = 391) were born between 340/7 and 366/7 weeks GA. Mean ± SD GA at birth among term-born children was 39.2 ± 1.5 weeks.Of 4287 births in the cohort, data from 4107 children were eligible for inclusion in this study. Children born preterm  shifted the distributions of LAZ and WAZ higher and resulted in significantly higher mean At birth, using CA  attenuated the overall prevalence of stunting 8.3 vs 11.6%) and underweight (3.7 vs 6.9%)  as opposed to age from conception . The extent to which anthropometric 0 to 18% —the misc0 to 18% –15 may oz scores to preterm-born children, (2) reduced variance in the distribution of z scores leading to greater precision in the standard errors, and (3) improved normality of the z score distributions during early infancy.Although the importance of using CA in analyses of anthropometric data is most apparent in early infancy , there are no known disadvantages of using CA up to 24 months of age in epidemiologic analyses, as is commonly done in clinical practice. In fact, using a GA-specific age scale in the application of growth standards confers several analytical advantages, including: (1) lower frequency of “biologically-implausible” outliers  , which mall children in sensitivity analyses) at the 3-month and later visits were assessed using only WHO-GS, and therefore demonstrated that observed differences in CA versus postnatal age estimates were not an artifact of differences in standards but a true effect of accounting for GA in the standardization of anthropometric indices and indicators. Nonetheless, further research is needed to fully explore the implications of combining multiple different standards for longitudinal assessment of child growth trajectories.An important limitation of this study was the use of a derived GA variable based on LMP and Dubowitz methods. These methods for GA assessment are less precise in comparison to first or second trimester ultrasound assessment. To minimize the risk of measurement error, we excluded all observations with implausible birthweight and GA combinations, which may have resulted in some selection bias by excluding observations at extreme ends of the birthweight-for-GA distribution. However, given that available ultrasound measurements in this cohort were prone to selection bias for preterm births, the systematically derived GA estimate provided a more reliable measure of GA for this study. Importantly, inferences regarding the effect of using CA compared to postnatal age for preterm-born children in the application of growth standards are robust against GA estimation methods. The effect of using CA for children born preterm in the application of WHO-GS in this study was likely a conservative estimate given the lower prevalence of undernutrition in Pelotas, Brazil, compared to countries in Sub-Saharan Africa and South Asia . In addiIn summary, this study suggests that the choice of age scale used to standardize anthropometric measures has substantial and meaningful implications for population-level epidemiologic inferences related to patterns of growth faltering and risk factors that contribute to undernutrition in early life. In LMICs with concurrently high prevalence of undernutrition and high incidence of preterm births, the discrepancies between postnatal age versus CA estimates of undernutrition at the population-level would be expected to be greater than the effect observed in this study, and may potentially misguide public health intervention priorities to reduce the population burden of undernutrition Future epidemiologic studies should therefore collect high-quality information about GA whenever possible and incorporate the GA information in the analyses of anthropometric data, at least at birth and during infancy. Longitudinal studies that account for heterogeneity in GA at birth in the application of growth standards are needed to better understand postnatal growth trajectories among children born preterm, the contribution of preterm birth to undernutrition in early life, and the implications for health outcomes in later life.


Birth centres are described as settings where women with uncomplicated pregnancies can give birth in a home-like environment assisted by midwives and maternity care assistants. If complications arise or threaten, the woman is referred to a maternity unit of a hospital where an obstetrician will take over responsibility. In the last decade, a number of new birth centres have been established in the Netherlands, based on the assumption that birth centres provide better quality of care since they offer a better opportunity for more integrated care than the existing system with independent primary and secondary care providers. At present, there is no evidence for this assumption. The Dutch Birth Centre Study is designed to present evidence-based recommendations for organization and functioning of future birth centres in the Netherlands. A necessary first step in this evaluation is the development of indicators for measuring the quality of the care delivered in birth centres in the Netherlands. The aim of this study is to identify a comprehensive set of structure and process indicators to assess quality of birth centre care.We used mixed methods to develop a set of structure and process quality indicators for evaluating birth centre care. Beginning with a literature review, we developed an exhaustive list of determinants. We then used a Delphi study to narrow this list, calling on experts to rate the determinants for relevance and feasibility. A multidisciplinary expert panel of 63 experts, directly or indirectly involved with birth centre care, was invited to participate.A panel of 42 experts completed two Delphi rounds rating determinants of the quality of birth centre care based on their relevance (to the setting) and feasibility (of use). A set of 30 determinants for structure and process quality indicators was identified to assess the quality of birth centre care in the Netherlands.We identified 30 determinants for structure and process quality indicators concerning birth centre care. This set will be validated during the evaluation of birth centres in the Dutch Birth Centre Study. Internationally, birth centres are described as settings where women with uncomplicated pregnancies can give birth in a home-like environment. In the Netherlands, women with uncomplicated pregnancies can choose where they want to give birth: at home, in a birth centre or in a hospital . At any Birth centres have been present in the Netherlands since the nineteenth century , but notIn recent years perceptions about the safety of the maternity care system in the Netherlands began to change. An important cause for this was the publication of the Euro-Peristat data, alarming the Netherlands because of its relatively high perinatal mortality compared to other European countries . It was quality of care is the degree to which health services for individuals and populations increase the likelihood of desired health outcomes and are consistent with current professional knowledge” [Although formulated in 1990, the definition of quality of care provided by the Institute of Medicine (IOM) is still widely accepted: “owledge” . Usuallyowledge” . Measureowledge” . Internaowledge” , 19, butowledge” –27, but In this article we describe a set of determinants for structure and process indicators for assessing the quality of birth centre care and we explain the approach we used to develop this set. We only describe the development of determinants for structure and process indicators, because a newly validated Optimality Index (OI-NL2015) and a Composite Adverse Outcome Score (CAOS) were used to evaluate outcomes of birth centre care .Identification of existing quality indicators in birth careIn order to develop a comprehensive set of structure and process quality indicators to evaluate birth centre care, we used mixed methods. Three phases were followed in the development process: 1) identification of existing quality indicators in birth care, 2) translating these structure and process indicators into determinants, 3) determinant selection by Delphi consultation. The study was conducted in the first half of 2013 as part of the Dutch Birth Centre Study [2.Translating indicators into determinantsIn the first phase of the study, we used various sources to find existing quality indicators in birth care. We began with an Internet search for documents from Dutch Institutes that had developed quality indicators for maternity care. Documents that described the (development of) quality indicators by midwives, obstetricians and maternity care assistants were obtained. Next, we reviewed international scientific literature about birth centres in order to identify existing quality indicators. We searched PubMed and the Cochrane Library using the Mesh terms: “birthing centres”, “quality indicator”, “health care” and search terms “quality” and “birth centre”. We used references from these articles to find other relevant articles and documents related to quality indicators in maternity care.3.Determinant selection by Delphi consultationIn the second phase we translated the structure and process indicators that we had identified into determinants (or topics): elements that identify the nature of the indicator. We used a framework based on the seven domains of quality according to the IOM  We added an eighth domain, “Law on the Accessibility of Healthcare Facilities”, because of obligations placed on healthcare facilities by this law in the Netherlands. The research group used their experience to add topics that were missing in the resulting list. No outcome indicators were included. We then created a questionnaire that members of an expert panel could complete in a minimum amount of time in order to maximize our response rate.We initiated an online Delphi study with the goal of obtaining consensus among a group of experts. The online Delphi technique is an anonymously structured approach, in which information is gathered from a group of participants through a number of Delphi rounds. The web-based anonymous nature of the Delphi technique ensures that a single individual cannot dominate the consensus formation. Moreover all participants are equally able to change their opinion in the course of the process , 29. OurWe selected participants for the expert panel from the Research Advisory Group of the Dutch Birth Centre Study , particiIn May 2013, we sent a link to an online questionnaire by e-mail to the expert panel. The experts were instructed to rate the determinants on relevance (to the setting) and feasibility (of use) and, if necessary, to comment on them. Each determinant was rated on a seven-point Likert scale . Finally, experts were encouraged to suggest additional relevant subjects that should be taken into consideration in the assessment of the quality of birth centre care. All ratings from the first Delphi round were analysed in Excel and distributions of scores were presented in median scores for each determinant. We considered determinants with a median score of ≥6 with agreement to be relevant and feasible to collect and accepted these immediately. Agreement was defined when 80% or more of the ratings were within a range of three (i.e. 5–6-7 of 4–5-6). Determinants that scored with a median score of ≤2 were rejected. Median scores of >3 and <6 with agreement or ≥6 without agreement were discussed again in the second Delphi round. Furthermore, all the comments on determinants from the first round were analysed and the descriptions of determinants were re-phrased in cases of ambiguity. All proposed new determinants from the first round were categorized in domains. New determinants were coded and two researchers of our research group decided, using a consensus method, which determinants should be submitted in the second round. Items the research group already had decided to include in the overall study  were not included in the second round.In the second Delphi round, the experts were informed about the median scores on relevance and feasibility of the total expert group, their own scores and the comments of the respondents regarding determinants for which no consensus was reached in the first round. They were instructed to re-consider their rating of the determinants presented in the first round as well as to rate and comment on the new elements the same way as in the first round. This was done to allow experts to revise their opinion of the first round while considering the ratings and comments provided by the other members of the expert panel. The link to the personalized online questionnaire was sent by email 10 days after the first round. Again, the median scores and the degree of agreement were calculated. Only scores ≥6 with agreement were adopted into the list. Determinants with scores for relevance ≥6 with agreement, but feasibility between 3 and 6 were presented to the research group for a final decision.Figure Two hundred fifteen indicators were derived from Dutch sources, 145 from international literature. We eliminated duplication and excluded all outcome indicators. Indicators that clearly do not determine quality of care in a birth centre  were also excluded from this list. Finally 66 structure and process indicators were identified.These 66 literature-based indicators were divided into seven themes matching the seven domains of quality according to the United States Institute of Medicine (IOM). The research group added a domain “Law Accessibility of Healthcare Facilities”. In these eight domains, 22 topics were identified, because several indicators appeared to relate to the same topic, albeit with different wordings. The research group added another 13 topics that they missed, based on their experience. After this process, the topics were formulated as 35 determinants to be included in the first questionnaire for the online Delphi panel.The questionnaire in the first Delphi round was completed by 48 experts (response rate of 76%). 42 of them also completed the questionnaire in the second round (response rate of 88%). During the first round, 24 of the 35 determinants were accepted for inclusion, none were rejected right away, leaving 11 topics without consensus. 22 experts mentioned 52 new topics they missed in the questionnaire. These topics were labelled and categorized, after which two researchers of our research group decided, based on consensus, that 8 of them would be included in the second Delphi round. In the second round, the 11 topics from the first round on which no consensus was reached and the 8 new topics were presented to the expert panel. This resulted in the acceptance of another five determinants and the rejection of 13 determinants. One determinant was presented to the research group because of low feasibility according to the experts. The research group accepted this determinant, so finally 30 determinants resulted from the Delphi consultation. Table In this study, part of the Dutch Birth Centre Study, we identified a set of 30 determinants, to be translated into 30 structure and process quality indicators that can be used to assess the quality of birth centre care in the Netherlands. The new developed determinants are derived from existing quality indicators in maternity care in the Netherlands  and indicators derived from international documents concerning birth centre care. The experts selected 5 determinants that are used by Laws in the research on characteristics and practices of birth centres in Australia  and 4 deA strength of the development of this set of determinants for indicators is that it is developed in collaboration with all parties involved in birth centre care, and is based on consensus. Therefore it can be expected that all professionals in the field will accept assessing the quality of birth centre care using this set of indicators.We are aware that the set of determinants for indicators we developed has its limitations. Firstly, to assess care in general, structure, process and outcome indicators should be used. However, because there are already a large number of quality indicators to assess outcomes of birth centre care, this set contains only structure and process indicators . The expFinally, although our study was focused on Dutch birth centres, we expect that this set of determinants for indicators will be applicable in other settings where birth centres are used.We used an online Delphi-method to develop a list of thirty determinants for structure and process indicators to measure quality of birth centre care. We will describe the process for developing quality indicators from these determinants and evaluation of the validity and reliability of these indicators as part of the Dutch Birth Centre Study in a later paper. It is important to underscore that indicators are part of an on-going cycle of quality improvement. Indicators should never be static. Changes in evidence or clinical relevance, a consistently high performance or a low variation in achievement, new developments and demographic changes in the population of childbearing women, all may be criteria for removing an indicator or adding a new one in a future list of determinants for quality indicators for birth centre care.


Preventive strategies of congenital anomalies are basically relying on the systematic ongoing collection and analysis of data and timely dissemination of information. The aim of this paper is to briefly report a critical review of a surveillance system of congenital anomalies in a developing country, by describing the challenges and experience of the registry since it began. Tabriz Registry of Congenital Anomalies (TRoCA) was mainly set up based on the guidelines provided by the International Clearinghouse for Birth Defects Surveillance and Research (ICBDSR) for data collection, coding, process, analysis, use, and evaluation of the system. TRoCA has successfully achieved its main objective as a pilot model for setting up a nationwide registry of congenital anomalies in the country. The programme has too succeeded in relation to its regional objectives: epidemiological rates and data have been produced consistently for etiological investigations, methodological studies, service provision, and preventive measures for selected anomalies. Our successful experience, as a small registry in a developing country, might be of interest and useful to practitioners, policymakers of birth defects control programmes, and mainly those willing to set up a monitoring system of congenital anomalies in similar areas. Birth defects are making a proportionally major contribution to perinatal mortality, childhood morbidity, and disability in many countries. Occurrence of congenital disorders varies between different countries ranging from 2 to 10 percent of births [The prevention of congenital anomalies requires prior knowledge of the aetiology and causal factors involved. Although aetiology is still largely unknown, preventive methods are now available for about 60 percent of congenital abnormalities , 4. PrevThe aim of this paper is to briefly report a critical review of a monitoring system of congenital anomalies in Iran, by describing the challenges and experience of the registry since it began.In 2000, a project was carried out in the Tabriz city of Iran to investigate the epidemiology of congenital anomalies. The aim of the study was to provide baseline information to set up a regional registry of birth defects for the first time in the country. This programme was then called Tabriz Registry of Congenital Anomalies (TRoCA) . Our proSome of the registry systems of the ICBDSR and EUROCAT members were studied in terms of data collection, process, analysis, use, and evaluation of the system to determine the requirements for setting up a local registry in Iran. The minimum requirements for the registry were also determined.TRoCA is located in Tabriz city run under the Tabriz University of Medical Sciences. Tabriz is one of the three major cities in Iran, located in the northwest region. Tabriz University of Medical Sciences is one of the five top universities in the country providing medical and health services for the population in the northwest of Iran. TRoCA has been financially supported by local and national funds.register the occurrence of selected birth defects in the region,prepare epidemiological indexes to indicate the magnitude and trends over time,monitor emerging or unusually high occurrences of congenital anomalies,make valid data available to policymakers,plan and implement preventive and control strategies to prevent selected anomalies,evaluate prevention and control strategies.The principal aims of TRoCA programme are to establish a monitoring system of congenital anomalies in the region and to implement control and preventive tasks in the area. It was primarily intended to use TRoCA framework as a pilot model for setting up a surveillance of birth defects in the whole country. Purposes of TRoCA are toPrenatal care is routinely provided for every pregnant woman on a regular basis (up to eight times) with 1–3 diagnostic sonographies during pregnancy. If needed, further diagnostic procedures  are performed. Termination of pregnancies is permitted for a few selected anomalies. TRoCA reports termination rate for major malformations only. After birth, all children in three hospitals involved in the programme are normally examined by a gynaecologist, obstetrician, neonatologist, or pediatrician at birth. They are followed up until hospital discharge for general health status, maturity, and congenital anomalies. The TRoCA programme covers about 15,000 births  in the area with about 300–400 newborns with one of the anomalies in this population. Background information and basic characteristic data are gathered for all births in TRoCA region. Some additional information  is also available for infants with anomaly and mothers. Karyotype and autopsy are not routinely performed unless it is requested as a necessity after full clinical investigation.Nervous system anomalies;Genitourinary tract and kidney;Anomalies of limb;Chromosomal anomalies;Cleft lip with/without palate;Congenital heart disease;Musculoskeletal and connective tissue anomalies;Digestive system anomalies;Eye and ear anomalies;Other anomalies.We use a “passive” method of data collection. The responsible persons (as registrar) for data documentation are midwives. A medical coder has been assigned in this programme to code/classify the defects. The end users defined the congenital anomalies for the purposes of this programme based on the standard coding system of the International Classification of Diseases (ICD) under one of the following main headings according to the primary diagnosis of anomaly: Total prevalence is calculated by dividing the numerator  by the relevant denominator  for the same period of time. An infant/fetus with more than one anomaly is counted once only in the numerator. The main criteria of inclusion of fetal deaths or stillbirths in data analysis are pathologic confirmation of the defect provided by the hospitals. Time trend analysis, relative frequencies, and confidence intervals are also calculated for some statistical purposes. For more details of the methodology, TRoCA publications may be searched.TRoCA activities have been approved by the Ethics Committee of the Tabriz University of Medical Sciences. Confidentiality and privacy of identity-related information are strictly considered in every part of the data gathering, handling, processing, registration, access, and reports.To date, generally, our data shows no epidemic of any of kinds of birth defects in the area. However, an unusual increase occurred for the total prevalence of congenital anomalies in 2002 in the registry region. We then identified that this happened due to an improvement in case ascertainment.As seen in A total of 261 024 births were registered in the region over the study period including 258 153 (98.9%) live births and 2871 (1.1%) stillbirths. During this period, 5870 cases with a primary diagnosis of congenital anomaly were ascertained, representing an overall prevalence rate of 224.9 per 10,000 births. Genitourinary tract and kidney anomalies, limb defects, anomalies of nervous system, and congenital heart diseases accounted proportionally for more than 65 percent of anomalies in the region .Data provided by TRoCA have resulted in a study to estimate the missing frequency of congenital cardiac anomalies at the time of delivery and birth in the region. Accordingly, 59.1% of children with congenital heart diseases were not identified at birth .We found that the accuracy of family physicians in case detection and diagnosis of congenital anomalies in rural areas is more than 98% , and assIn addition to the registration of birth defects, TRoCA has extended its activities to implement control and preventive services for genetic disorders and congenital anomalies in the region. This includes genetic services to families who have a history of an anomaly/disorder in the family, and preconceptional programmes for young couples. This new programme called Tabriz Foundation for Public Health Genetics (TFPHG) was launched in 2013 .General information is routinely collected for every neonate. Some exposure information is also available of mothers of all malformed infants. Other women giving births in the TRoCA maternity and children hospitals with normal newborns routinely complete the similar data form. They might be considered as matched control group. Using these data of control group plus routine statistics from general population , testingTRoCA is always able to describe the distribution and occurrence of congenital anomalies in its defined population . However, epidemiological reliability and representativeness of the rates captured by the programme have not been fully investigated yet. The timeliness of the information provided by TRoCA is also still a matter of epidemiological pitfall where the programme is able to release primary information on the occurrence of congenital anomalies at least one year after the data collection.As indicated before, TRoCA programme monitors about 15,000 births  in the area with about 300–400 cases with one of the defects in the newborns. The very small frequency and the rare nature of some groups of anomalies influence the epidemiological power of the rates and occurrence patterns of various types of congenital anomalies provided by the programme.Multiple sources of case ascertainment result in a true pattern of birth defects in the population. The data of TRoCA comes mainly from two maternity hospitals plus a referral medical centre for sick children in the region. Some anomalies are routinely not diagnosed until some times after birth. For this reason, ascertainment of those defects is inevitably delayed. Nearly 60% of children born with congenital heart disease, for example, were not ascertained by TRoCA at birth. They were then identified as having heart defects in children hospital during the next 12 months after birth . The delAt birth, nervous system anomalies were ranked first in the TRoCA earlier years' data while the first rank belongs to congenital heart defects in recent years' records . The reaTabriz Registry of Congenital Anomalies started its main activities in 2000. A nationwide registry of congenital anomalies in Iran was then established in 2012 based on the data, framework, and baseline structure provided by TRoCA (as a pilot programme). It is therefore believed that TRoCA has successfully achieved its main objective as a pilot model for the whole country.TRoCA has too seemingly succeeded in relation to its regional objectives: annual prevalence rates have been produced consistently over years , geograpTRoCA recorded a low rate for spina bifida, high rates for anencephaly, hydrocephaly, and cleft palate without cleft lip, and a very high rate of limb reduction defects in the region. While TRoCA examines and reports the annual rates of congenital anomalies consistently, we do not know for sure whether any change in these rates is due to a true existence/absence of epidemics in the region, due to the technical failure of our monitoring, or due to aetiologic and environmental teratogens. We do not have any parallel surveillance for environmental teratogens. The reasons behind the high/low rates of some groups of anomalies are still unclear. There need therefore for further investigations for every selected anomaly with unusual rate of occurrence in the area.As a programme in a developing country, the condition of antenatal screening procedures, detection and ascertainment methods of defects, variable validity of the data, low coverage and small denominator population (due to the limited sources of funding), small number of cases , low power of the programme to find rare defects, and no full follow-up plan for anomalies after hospital discharge are still among the main limitations of the registry. The lack of epidemiological power and representativeness should therefore be carefully considered in examining the rates and patterns of various types of congenital anomalies, and when investigating for possible local causes and influencing factors of birth defects in the area.For evaluation purposes, we assess the strengths and weaknesses of the TRoCA programme based on the major evaluation components of a standard monitoring system .TRoCA has, so far, relied on staff for whom this programme was not their core duties. Allocating fully funded full time individuals for TRoCA may help to solve some parts of the above problems in its future activities.It is concluded that as a small registry of congenital anomalies in a developing country, although TRoCA is still facing some challenges and problems, we believe that it has been successful in achieving its main objectives. Our experiences might be of interest and useful to practitioners, policymakers of birth defects control programmes, and mainly those willing to set up a monitoring system of congenital anomalies.


In utero exposure to infections is associated with adverse neurocognitive outcomes in the offspring. Elevated maternal prenatal serum inflammatory markers, such as C-reactive protein (CRP), have been associated with increased risks of neurodevelopmental disorders, including schizophrenia, later in life. The objective of this study is to investigate the associations between elevated serum concentrations of CRP in early gestation, prospectively assayed in maternal sera, and adolescent psychotic experiences and academic performance. We hypothesised that elevated maternal CRP is associated with adolescent psychotic experiences and poorer academic performance.Using data from the Northern Finland Birth Cohort 1986 (NFBC1986), a prospective birth cohort including data since before birth, we examined the association between maternal CRP levels in early gestation  and adolescent psychotic experiences  and poorer academic performance , controlling for sex and maternal education level using multivariable regression analysis. Prior to analyses we determined there was sufficient power to detect small associations (OR>1.68) for these variables.After controlling for sex and maternal education, those in the highest tertile of prenatal maternal CRP had increased odds of auditory hallucinations (on the PROD-screen) at age 16 years , and poorer school academic performance .Maternal prenatal immune activation is associated with neurodevelopmental outcomes in adolescent offspring. This work extends previous findings regarding prenatal/childhood immune activation and clinical psychiatric diagnoses in adult offspring.


EFW) and fetal biometry are complementary measures used to screen for fetal growth disturbances. Our aim was to provide international EFW standards to complement the INTERGROWTH‐21st Fetal Growth Standards that are available for use worldwide.Estimated fetal weight (FGLS) and INTERBIO‐21st Fetal Study (FS), two components of the INTERGROWTH‐21st Project, had ultrasound scans every 5 weeks from 9–14 weeks' until 40 weeks' gestation. At each visit, measurements of fetal head circumference (HC), biparietal diameter, occipitofrontal diameter, abdominal circumference (AC) and femur length (FL) were obtained blindly by dedicated research sonographers using standardized methods and identical ultrasound machines. Birth weight was measured within 12 h of delivery by dedicated research anthropometrists using standardized methods and identical electronic scales. Live babies without any congenital abnormality, who were born within 14 days of the last ultrasound scan, were selected for inclusion. As most births occurred at around 40 weeks' gestation, we constructed a bootstrap model selection and estimation procedure based on resampling of the complete dataset under an approximately uniform distribution of birth weight, thus enriching the sample size at extremes of fetal sizes, to achieve consistent estimates across the full range of fetal weight. We constructed reference centiles using second‐degree fractional polynomial models.Women with an accurate gestational‐age assessment, who were enrolled in the prospective, international, multicenter, population‐based Fetal Growth Longitudinal Study (AC and HC (without FL) as log(EFW) = 5.084820 − 54.06633 × (AC/100)3 − 95.80076 × (AC/100)3 × log(AC/100) + 3.136370 × (HC/100), where EFW is in g and AC and HC are in cm. All other measures, gestational age, symphysis–fundus height, amniotic fluid indices and interactions between biometric measures and gestational age, were not retained in the selection process because they did not improve the prediction of EFW. Applying the formula to FGLS biometric data (n = 4231) enabled gestational age‐specific EFW tables to be constructed. At term, the EFW centiles matched those of the INTERGROWTH‐21st Newborn Size Standards but, at < 37 weeks' gestation, the EFW centiles were, as expected, higher than those of babies born preterm. Comparing EFW cross‐sectional values with the INTERGROWTH‐21st Preterm Postnatal Growth Standards confirmed that preterm postnatal growth is a different biological process from intrauterine growth.Of the overall population, 2404 babies were born within 14 days of the last ultrasound scan. Mean time between the last scan and birth was 7.7  days and was uniformly distributed. Birth weight was best estimated as a function of EFW, as an adjunct to routine ultrasound biometry, from 22 to 40 weeks' gestation. However, we strongly encourage clinicians to evaluate fetal growth using separate biometric measures such as HC and AC, as well as EFW, to avoid the minimalist approach of focusing on a single value. © 2016 Authors. Ultrasound in Obstetrics & Gynecology published by John Wiley & Sons Ltd on behalf of International Society of Ultrasound in Obstetrics and Gynecology.We provide an assessment of  One of the main objectives of antenatal care is screening for fetal growth disturbancesUltrasound evaluation of the fetus involves measuring head circumference (HC), abdominal circumference (AC) and femur length (FL), and the values can be combined to calculate an estimated fetal weight (EFW); this estimate is often used alone in clinical practice without considering the individual measurements. However, we believe that arguments concerning the most appropriate single parameter to use are inappropriate because clinicians should use all the tools available in their armamentarium for making crucial clinical decisions that have major implications for both mothers and newborns.st standards for HC, AC and FL, size at birth and postnatal growth in preterm infantsst Project were: (1) to develop a formula to estimate fetal weight based on ultrasound biometry and birth weight; and (2) to construct international EFW standards for fetuses at 22 to 40 weeks' gestation.The development of international EFW standards is overdue, and these should share the same conceptual basis as the published INTERGROWTH‐21st is an international, multicenter, population‐based project consisting of a number of components, including the Fetal Growth Longitudinal Study (FGLS) and INTERBIO‐21st Fetal Study (FS).INTERGROWTH‐21FGLS was conducted between 27 April 2009 and 2 March 2014 in eight urban areas: the cities of Pelotas (Brazil), Turin , Muscat (Oman), Oxford (UK) and Seattle (USA); the Shunyi County of Beijing (China); the central area of Nagpur (India); and the Parklands suburb of Nairobi (Kenya). The primary aim was to study longitudinally the health and development of fetuses into infancy, by monitoring growth, health, nutrition and neurodevelopment from less than 14 weeks' gestation to 2 years of age, so as to produce prescriptive growth standards to complement the existing WHO Child Growth Standards. This was achieved by studying a cohort of healthy, well‐nourished, pregnant women who were at low risk of adverse maternal and perinatal outcomes at both population and individual levels. The study details have been described elsewhereIn contrast, FS recruited an unselected cohort of pregnant women, between 8 February 2012 and 24 December 2015, from three FGLS sites , and three new sites . The primary aim was to study the effects of various intrauterine exposures  on growth, health, nutrition, neurodevelopment and the epigenome, over the same developmental age range, i.e. from less than 14 weeks' gestation to 2 years of age.To develop the EFW formula requires as many pregnancies as possible that have a standardized scan and birth‐weight measurement. In order to achieve this we included fetuses from both FGLS and FS; only those that had an ultrasound scan within 14 days of birth were included in the calculations. To develop the international standards for EFW, the formula derived was then applied to the healthy FGLS population from which the International Fetal Growth Standards were producedst Project was approved by the Oxfordshire Research Ethics Committee ‘C’ (reference: 08/H0606/139), the research ethics committees of the individual participating institutions and the corresponding regional health authorities in which the project was implemented. Participants provided written consent to be involved in the study.The INTERGROWTH‐21In both studies women were recruited at less than 14 weeks' gestation. All women underwent ultrasound measurement of fetal crown–rump length (CRL) using standardized methodologyst website.Following the dating scan, women were scanned every 5 weeks (±1 week), so that the possible ranges were 14–18, 19–23, 24–28, 29–33, 34–38 and 39–42 weeks' gestation. At each visit, fetal HC, biparietal diameter (BPD), occipitofrontal diameter (OFD), AC and FL were measured three times from three separately obtained ultrasound images of each structure. The detailed measurement protocols, including graphical displays of measurement techniques, and the unique standardization procedures for all measurements and sonographer training have been reported elsewhereBriefly, head measurements were taken in an axial view at the level of the thalami, with an angle of insonation as close as possible to 90°. The head had to be oval in shape, symmetrical, centrally positioned and filling at least 30% of the monitor screen. The midline echo  had to be broken anteriorly, at a third of its length, by the cavum septi pellucidi. The thalami had to be located symmetrically on either side of the midline. Calipers were then placed on the outer border of the parietal bones (outer to outer) at the widest or longest part of the skull for the BPD and OFD, respectively; HC was measured using the ellipse facility on the outer border of the skull.AC measurements were taken in a cross‐sectional view of the fetal abdomen as close as possible to circular in shape, with the umbilical vein in the anterior third , with the stomach bubble visible. The sonographer was instructed to avoid applying too much pressure with the transducer, which can distort the circular shape of the fetal abdomen. The abdomen had to fill at least 30% of the monitor screen, and the spine had to be at either the 3 or 9 o'clock position to avoid internal shadowing; the kidneys and bladder had not to be visible. For the measurements, the contour of the ellipse was placed on the outer border of the abdomen.st website).Finally, FL was measured using a longitudinal view of the fetal thigh closest to the probe and with the femur as close as possible to the horizontal plane. The angle of insonation of the ultrasound beam was about 90°, with the full length of the bone visualized, unobscured by shadowing from adjacent bony parts, and the femur had to fill at least 30% of the monitor screen. The intersection of the calipers was placed on the outer borders of the edges of the femoral diaphysis (outer to outer) ensuring clear femoral edges; ultrasound artifacts of the femoral edges such as the proximal trochanter or pointed femoral spurs were not included in the measurement , was used at all sites. To avoid expected value bias, the machine was adapted so that fetal measurements were not visible to the sonographer on the screen. Only after three measurements of each structure had been recorded were the average values revealed for clinical purposes. All ultrasound data were submitted electronically to the study database. Data were entered locally directly onto the web‐based systemAfter taking each set of measurements, sonographers scored the quality of their images on the basis of standard image‐scoring criteriaBirth weight was measured within 12 h of birth using identical electronic scales  at all sites. The equipment, which was calibrated twice a week, was selected for accuracy, precision and robustness, as shown previouslyZ‐scores using the INTERGROWTH‐21st equationsHC, BPD, OFD, AC and FL, in cm or transformed into gestational age on the day of the last scan, in weeks;symphysis–fundus height, in cm;amniotic fluid, assessed by the deepest vertical pool and amniotic fluid index in cm;2.cross‐sectional head area and abdominal area computed from their orthogonal diameters, in cmFrom the FGLS and FS cohorts, we identified all live babies without any congenital abnormality who were born at > 24 weeks' gestation and within 14 days of the last ultrasound scan. Given the study design, we expected the births to have occurred uniformly between 0 and 14 days after the last ultrasound scan, i.e. we expected there to be a mean time of 7–8 days between the last scan and birth. This cut‐off allowed a greater number of births at low gestational ages to be included, for which most of the existing formulae have been prone to prediction error, probably because scant data exist for estimationWe hypothesized that the contribution of any given anthropometric measurement to EFW might vary with gestational age. Therefore, we also considered interactions between HC, BPD, OFD, AC and FL and gestational age on the day of the last scan. Statistical modeling was conducted using second‐degree fractional polynomialsSome prediction bias would be expected because of significant growth between the day of the last scan and birthAs expected, owing to the prospective, population‐based design of FGLS, most births occurred close to 40 weeks' gestation, meaning that the scatter of observations across the 22–40‐week window was very uneven. We were aware that estimation using the complete dataset would yield very accurate estimates at 40 weeks' gestation, where the greatest contribution of the data is found, but with limited model validity for lower birth weights. To overcome this problem and allow accurate birth‐weight estimation over the whole range of observed data, we constructed a bootstrap model selection and estimation procedure based on resampling of the complete dataset under an approximately uniform distribution of birth weightvs fitted plots. Given that we estimated fetal weight at the time of the last scan using an average model for growth, we investigated the bias of our model for EFW by calculating the mean of percent prediction errors defined by the formula (100 × (EFW − birth weight)/birth weight), for decreasing time‐to‐birth intervals (i.e. from 14 to 0 days). Finally, we also calculated the absolute percent prediction error defined by the mean of the absolute prediction errors.Assessment of goodness of fit in the complete dataset relied on inspection of residuals with quantile–quantile (q‐q) plots and residuals rd or 97th centile, and regression‐based reference limitsrd or 97th centile. Further details on the precision obtained at the 5th or 10th centile by sample size (ranging from 500–6000) have been included in a table in a previous publicationThe construction of reference centiles was based solely on FGLS data. The sample size was based on pragmatic and statistical considerations; the latter focused on the precision and accuracy of one extreme centile, i.e. the 3The data from all the study sites were pooled to construct the Fetal Growth StandardsWe explored the following statistical methods: mean and SD method using fractional polynomialsUsing de‐trended q‐q plots (worm plots), significant evidence of deviations from normality was seen so we resorted to using the more complex LMS, LMST and LMSP methods allowing for skewness and kurtosisrd and 97th centiles) over the whole 22–40 weeks' gestational‐age range. Second, we considered mixed‐effect models accounting for repeated measurements within LMS, LMST and LMSP frameworks. This analysis also showed no impact on the estimated centiles. The best fit was found using a three‐parameter Box–Cox Gaussian distribution (i.e. the LMS method) for the response variable with a second‐degree fractional polynomial functional form for gestational age. This method also gives estimated SDs of EFW, allowing estimation of centiles.As most of the women had four to six ultrasound scans, the effect of correlated data within fetuses was investigated. First, in a sensitivity analysis, a random observation time was sampled for each fetus and the modeled centiles in this subset were compared visually with the complete dataset. The approach is justified by the experimental design of the study, which ensures non‐informative observation timesvs fitted values.Goodness of fit for the overall model was assessed by comparing empirical centiles  with the fitted centiles, using de‐trended q‐q plots of the residuals across gestational ageAll analyses were carried out in R statistical softwaren = 1556) and FS (n = 848) who were born within 14 days of the last ultrasound scan were examined , the actual fetal weight at the time of the last scan was best estimated as a function of AC and HC with the following formula:None of the other covariates including FL, BPD, OFD, gestational age, symphysis–fundus height, amniotic fluid indices or interactions between biometric measurements and gestational age was retained in the selection process. This model suggests a linear relationship between log(EFW) and HC. Despite the negative coefficients, the two terms involving AC describe an increasing sigmoid‐shaped relationship between AC and birth weight  for a fixed HC value of 26 cm  and −0.8%  in those born within 1 day (n = 198) , showing that our model was unbiased for predicting weight at the time of the last scan and that the correction we applied to compensate for time to birth was appropriate. In the group born within 1 day of the last scan, the mean absolute prediction error was 7.6%, with 80%, 90% and 95% of predicted weights falling within 11%, 14% and 18% of the true birth weight, respectively.The performance of the formula for EFW was assessed both by mean and absolute percent prediction errors; mean percent prediction error is used as a measure of potential bias of EFW due to growth between the last scan and birth, while mean absolute prediction error represents the dispersion of the errors. The mean percent prediction error steadily tended towards zero as the time interval between the last scan and birth decreased. Prediction error was −10.7%  in babies born exactly 14 days after the last scan  and observed values, assessed by gestational age‐specific comparisons, demonstrated excellent agreement. The overall differences between empirical and smoothed centiles were small, with mean ± SD differences of 16 ± 28 g, 13 ± 17 g and 5 ± 33 g for the 3rd, 50th and 97th centiles, respectively.Creation of the international EFW standards was based on the complete FGLS dataset. The gestational age‐specific observed and smoothed centiles for EFW are presented in Figure rd, 10th, 50th, 90th and 97th fitted centile curves for EFW according to gestational age, which represent the international standards, are presented in Figure Z‐scores. By estimating the EFW and knowing the gestational age, desired centiles can be calculated. For example, if AC = 26 cm and HC = 29 cm, at 30 + 0 weeks:The 3Z‐score at 30 weeks' gestation, using the equations in Table To compute the corresponding rd centile (α = 0.03), i.e. Z = −1.88 at 30 + 0 weeks, is calculated as follows using the equations in Table rd centile for EFW at 30 weeks' gestation is therefore: C0.03(30) = exp(7.008552) = 1106 g.Finally,rd, 10th, 50th, 90th and 97th centiles according to gestational age are presented in Table S1.The actual values for the 3st Project provides standards for early human growth based on populations that conform to the prescriptive approach recommended by the WHOst Project is unique because it has produced, for the first time, fetal ultrasound, newborn size and preterm postnatal growth datasets that have all been collected from the same underlying populations using the same rigorously applied methodologies.The INTERGROWTH‐21We now present international EFW standards to complement the existing set, along with a formula for EFW based on HC and AC. Compared with several previous formulaeUnusually, we lowered the starting gestational age to 22 weeks, 2 weeks below the customary cut‐off of 24 weeks' gestation for viability, for two reasons: to facilitate early recognition of fetal growth restriction around the recommended time of the second‐trimester anatomy scan and to anticipate a possible extension of the limit of viabilityst Newborn Size Standards at 40 weeksrd, 50th and 97th EFW centiles at 40 weeks are 2554 g, 3338 g and 4121 g, respectively (Table S1), whereas for newborns (sexes combined) they are 2591 g, 3321 g and 4154 g, respectively . These similarities between fetal‐ and birth‐weight centiles suggest that our model is valid for developing a formula for EFW using ultrasound biometry.At the upper end of gestation, the centiles closely match those of the INTERGROWTH‐21rd, 50th and 97th EFW centiles are 1495 g, 1954 g and 2529 g, respectively (Table S1); for newborns (sexes combined), they are 1190 g, 1903 g and 2715 g, respectively. It is possible that these differences are due to an overrepresentation of small, as well as, to a lesser extent, large babies in preterm births, even in the selected pregnant and newborn populations we studied.In contrast, there are significant discrepancies earlier in pregnancy Figure . For exaThe EFW formula and standards we present are also unique because we avoided the many common limitations identified by previous reviewsa‐priori strategyConversely, our standards are prescriptive, whereas reference charts describe only fetal size at a given place and time. The standards were derived prospectively, population‐based and multinational. We have shown  that the eight populations were consistently similar and could be pooled to create international standardsUsing ultrasound, we examined separately HC, AC and FL, providing a comprehensive evaluation of structures that have different growth patterns; these measurements are often combined to calculate EFW. There are advantages in using a summary approximation: it is the most commonly measured marker of size at birth; as birth weight is associated with morbidity and mortality, it is helpful when counseling parents and enables pediatricians to make management decisionsHowever, there are also disadvantages in using only a single summary measure of size: first, there is a loss of the most granular information available when using the individual measurements, in terms of fetal skeletal and fat‐based growth. Second, the fact that the individual measurement errors are compounded means that estimation is prone to inaccuracy; previous studies have shown that 95% prediction intervals for random error are in the region of ± 14% of birth weight, and this is a particular problem in low‐ and high‐birth‐weight babiesst standards describe optimal growth and can be used to assess both individuals and populations.Therefore, we strongly recommend that, for clinical use, all individual fetal measurements, together with the summary measure of EFW, should be used together to make clinical decisions. In perinatal medicine, there is no room for a quick, minimalist approach that might lead to the early delivery of an at‐risk fetus. Finally, implementation of the standards may raise concerns regarding the generalizability of data originating from a limited number of sites and/or a highly selected, low‐risk population. As we have argued previouslyAppendix S1 Members of the International Fetal and Newborn Growth Consortium for the 21st Century (INTERGROWTH‐21st and INTERBIO‐21st) and its CommitteesClick here for additional data file.Figure S1 Relationship between fetal weight and abdominal circumference in the final model, plotted for a fixed head circumference of 26 cm.Click here for additional data file.Figure S2 Relationship between fetal weight and head circumference in the final model, plotted for a fixed abdominal circumference of 23 cm.Click here for additional data file.Figure S3 Bias in estimation of fetal weight as a function of time to birth, showing mean percent prediction error and 95% CI according to time between last ultrasound scan and birth.Click here for additional data file.Figure S4 Gestational age‐specific centiles for estimated fetal weight (blue) and birth weight (red). 3rd, 10th, 50th, 90th and 97th centiles are shown.Click here for additional data file.Table S1 Estimated fetal weight per completed week of gestation at 3rd, 10th, 50th, 90th and 97th centilesClick here for additional data file.