Birth weight for gestational age is a widely-used proxy for fetal growth. Although the need for different standards for males and females is generally acknowledged, the physiologic vs pathologic nature of ethnic differences in fetal growth is hotly debated and remains unresolved.th percentile) small-for-gestational age (SGA) among live births based on both a single standard and four ethnic-specific standards.We used all stillbirth, live birth, and deterministically linked infant deaths in British Columbia from 1981 to 2000 to examine fetal growth and perinatal mortality in Chinese , South Asian , First Nations, i.e., North American Indian , and other  births. We used a new analytic approach based on total fetuses at risk to compare the four ethnic groups in perinatal mortality, mean birth weight, and "revealed" (< 10Despite their lower mean birth weights and higher SGA rates (when based on a single standard), Chinese and South Asian infants had lower perinatal mortality risks throughout gestation. The opposite pattern was observed for First Nations births: higher mean birth weights, lower revealed SGA rates, and higher perinatal mortality risks. When SGA was based on ethnic-specific standards, however, the pattern was concordant with that observed for perinatal mortality.The concordance of perinatal mortality and SGA rates when based on ethnic-specific standards, and their discordance when based on a single standard, strongly suggests that the observed ethnic differences in fetal growth are physiologic, rather than pathologic, and make a strong case for ethnic-specific standards. For exath percentile for gestational age according to two different standards: (1) the current British Columbia live birth standard [Gestational age-specific patterns of fetal growth restriction were estimated using an indirect method based on the fetuses-at-risk approach. The number of "revealed" (see below) small-for-gestational-age (SGA) live births was determined for each group based on a birth weight < 10standard  and (2) Because of the low absolute number of events  at early gestational ages for the Chinese, South Asian, and First Nations groups, we analyzed rates for these events as 2-week prospective risks. In other words, the rates were calculated as the number of events occurring during a given 2-week gestational period divided by the number of fetuses alive (and thus at risk for these events) at the beginning of that period. Neonatal deaths and stillbirths were analysed using the same method as perinatal deaths, with similar results (available on request).ratios of deaths to live births or total births at a given gestational age. They are not true proportions, because the denominator does not include all subjects (unborn fetuses) at risk for the events denoted by the numerator; all living fetuses are at risk for stillbirth, live birth, and early neonatal death in the succeeding week. Neonatalogists are (appropriately) concerned with mortality among live-born births at a given gestational age, but neither the pregnant woman carrying a live fetus at a given gestational age nor her obstetrician, family physician, or midwife has any way of knowing whether or not her fetus will be born in the next week. From the woman's and her unborn fetus's perspective, the risk of stillbirth or live birth and early neonatal mortality in the succeeding week does indeed increase with advancing gestation, because the likelihood of birth (either a live birth or a stillbirth) rises as gestation advances [We have previously shown that risks based on the number of fetuses at risk, rather than the number of total births, provides greater coherence between birth rates (and thus risks of early preterm birth), fetal growth restriction, and perinatal mortality ,21-23. Oth percentile birth at the given gestational age for this data set . Since the revealed SGA rate depends on both the birth rate and the SGA rate among live births, it is far below 10%, except in the last gestational age category (42+ weeks) when all remaining fetuses are born. It thus relates the number of live-born SGA infants to the number of fetuses at a given gestational age who were at risk for both SGA and birth during the subsequent week.Because SGA cannot be determined among unborn fetuses , and because the weight of stillbirths may underestimate the fetal weight at the (earlier) time of fetal death, we have developed a proxy measure, "revealed SGA," that provides a tip-of-the-iceberg indication of fetal growth restriction. The revealed SGA rate is the number of live-born SGA infants at a given gestational age divided by the number of fetuses at risk ,21-23, wWe used two different internal standards to define revealed SGA: (1) a single standard comprising all three study groups, and (2) a group-specific standard for each of the ethnic groups. We then graphically compared the patterns of gestational age-specific rates of live birth, revealed SGA, and perinatal death among the three study groups and compared the coherence of the patterns using the single vs group-specific SGA standards.rd order polynomial calculated as the least squares fit through data points according to the following equation: y = b + c1x + c2x2 + c3x3 where b and c are constants. Missing birth weight and gestational age values comprised <0.25% of total births in each group and were proportionally distributed across gestational age. Chinese and South Asians had the lowest percent missing, while First Nations was only slightly higher than Others. BCVSA makes a particular effort to include birth weight and gestational age values on all records. If either measure is not recorded on the notice of birth, the source is contacted before the record is processed.All statistical analyses were carried out using SAS-PC version 8.2. Specialized graphic output was produced using Microsoft Excel software Version 2002. Smoothing of the charts was accomplished using a 3The LBW rate  was 4.2% among Chinese, 6.3% among South Asian, 5.6% among First Nations, and 4.4% among Other births. The preterm birth (<37 completed weeks) rate was 5.3% among Chinese, 6.8% among South Asian, 9.3% among First Nations, and 5.6% among Other births. Table In contrast to the perinatal mortality curves, mean birth weights were consistently highest in First Nations births, whereas those for Chinese and South Asian births progressively lagged behind those of Other births after 35–36 weeks Figure . These dAs shown in Figure Revealed SGA rates based on ethnic-specific fetal growth standards Figure , howeverTable These data spanned a 20-year period and temporal trends in immigration and mortality might conceivably bias overall outcomes. To address this possibility, we re-analysed the birth and perinatal mortality data within 5-year periods from 1981 to 2000. A slight increase was observed in the proportion of births to First Nations and a slight decrease in the proportion born to the Other group, but perinatal mortality rates declined for both groups over the 20-year period. The proportion of Chinese and South Asian births increased substantially in the 1990s, while perinatal mortality rates rose slightly, although fluctuations due to low numbers resulted in some instability in the trends. A general convergence of mortality rates occurred, although the groups generally maintained their relative positions.We found lower mean birth weights and higher rates of revealed SGA among ethnic Chinese and South Asian births compared with other ethnic groups when the classification of SGA was based on a single British Columbia standard of birth weight for gestational age. First Nations births, on the other hand, had higher mean birth weights, and slightly lower revealed SGA rates near and after term, compared with other ethnic groups. These results are similar to those reported by previous investigators-10,13-15When ethnic-specific standards are used to define SGA, however, revealed SGA rates and perinatal mortality rates become far more concordant. To the extent that SGA prevalence among live-born infants reflects an adverse intrauterine environment , it shouOur finding that gestational age-specific patterns of revealed SGA cohere better with gestational age-specfic perinatal mortality when ethnic-specific standards were used suggests that ethnic-specific birth weight for gestational age represent a physiologic  rather than a pathologic process. If differences in fetal growth  were truly pathologic, rather than physiologic, we would expect patterns that were more coherent with those observed for perinatal mortality when the definition of SGA was based on a single population standard, rather than ethnic-specific standards.within the British Columbian population of births over a 20-year period a highly unlikely explanation for our findings. We are unaware of any selection factors that would simultaneously lead to smaller fetuses/infants and lower perinatal mortality.The fact that Chinese and South Asian births in British Columbia occur in immigrants (by definition--see Methods) raises the question of a "healthy migrant" bias and the generalizability of our findings to ethnic Chinese and South Asian births in China, South Asia, and elsewhere. Although it is possible that Chinese and South Asian parents who succeeded in immigrating to Canada are better educated and more socio-economically advantaged than those who remained in their countries of origin or those in the indigenous (non-immigrant) British Columbia population, the neighbourhood income distributions of the immigrant parents in our sample were lower than the income distribution in the province overall . All elsOur mortality rates in the South Asian group were lower than those in the Other group which contradicts the findings in some European studies ,25. Howein utero during the risk period  are at risk of stillbirth during the period but are not counted in the denominator at risk. With fetuses at risk in the denominator, however, the combination of stillbirths and early neonatal deaths as perinatal deaths is no longer problematic. Indeed, we have carried out separate analyses of stillbirths and neonatal deaths based on the fetuses-at-risk approach, and the results (available on request) are entirely consistent with those presented here for perinatal deaths.We have previously argued for separate analysis of stillbirths and early neonatal deaths and against their combination as "perinatal deaths" when gestational age-specific perinatal mortality risk is based on using total births (stillbirths plus live births) as the denominator for calculating the risk . Using tWhile factors other than ethnicity  can affect fetal growth and could confound the ethnic differences we observed; numerous previous studies-10,13-15The practical application of ethnic-specific standards require further "on-site" examination, but our results provide a useful analytic comparison tool for future research and should be considered when making clinical judgements about fetal surveillance, induction of labour, and postnatal nutrition.As in all studies based on birth and death registrations, errors can occur in the estimation of gestational age, the recording or computer entry of birth weight or registration, and the linkage of live births and infant deaths. The deterministic linkage method used in British Columbia should minimize the latter type of error, and detailed analysis of the quality of the birth weight and gestational age data from this file have been reported . NonetheThe concordance of perinatal mortality and SGA rates when based on ethnic-specific standards, and their discordance when based on a single standard, strongly suggests that the observed ethnic differences in fetal growth are physiologic, rather than pathologic, and make a strong case for ethnic-specific standards.The author(s) declare that they have no competing interests.WJK conceived the study, designed and performed all analyses (except Table The pre-publication history for this paper can be accessed here:


A rare case of lymphoepithelial cyst formed in the piriform sinus of the hypopharynx is reported. Histopathological examination revealed a lymphoepithelial cyst. It was removed by laryngomicrosurgical technique using a side-opened direct laryngoscope. Because this cyst was wide-based on the antero-medial region in the right piriform sinus of the hypopharynx, the mucous membrane around the cyst was incised electrosurgically and then detached to facilitate removal. In this paper, we describe our surgical procedure for removing the cyst in this case and discuss the possible causes of the disease.


The link between maternal factors and birth outcomes is well established. Substantial changes in society and medical care over time have influenced women's reproductive choices and health, subsequently affecting birth outcomes. The objective of this study was to describe temporal changes in key maternal and fetal factors affecting birth outcomes in Newcastle upon Tyne over three decades, 1961–1992.Particulate Matter and Perinatal Events Research (PAMPER) study. The PAMPER database was created using details from paper-based hospital delivery and neonatal records for all births during 1961–1992 to mothers resident in Newcastle . In addition to hospital records, we used other sources for data collection on births not included in the delivery and neonatal records, for death and stillbirth registrations and for validation.For these descriptive analyses we used data from a population-based birth record database constructed for the historical cohort The average family size decreased mainly due to a decline in the proportion of families with 3 or more children. The distribution of mean maternal ages in all and in primiparous women was lowest in the mid 1970s, corresponding to a peak in the proportion of teenage mothers. The proportion of older mothers declined until the late 1970s (from 16.5% to 3.4%) followed by a steady increase. Mean birthweight in all and term babies gradually increased from the mid 1970s. The increase in the percentage of preterm birth paralleled a two-fold increase in the percentage of caesarean section among preterm births during the last two decades. The gap between the most affluent and the most deprived groups of the population widened over the three decades.Key maternal and fetal factors affecting birth outcomes, such as maternal age, parity, socioeconomic status, birthweight and gestational age, changed substantially during the 32-year period, from 1961 to 1992. The availability of accurate gestational age is extremely important for correct interpretation of trends in birthweight. Maternal factors such as age and parity are known to influence birth outcomes. Thus advanced maternal age is associated with preterm birth -3, fetalWhile gestational age has been acknowledged as a major determinant of birthweight, it has not been collected as part of routine vital perinatal statistics in many countries, for example the UK . Even whParticulate Matter and Perinatal Events Research (PAMPER) study offers the unique opportunity to describe temporal changes in key maternal and fetal factors affecting birth outcomes in a single conurbation over three decades, from 1961 to 1992. More specifically, we describe trends in maternal age, parity, aggregate level socioeconomic status, birthweight and gestational age and also demonstrate a reduction in stillbirth and infant mortality by decade.The UK Newcastle upon Tyne, located within the Northern Region of England, has a current population of approximately 260,000 inhabitants. The population structure of the Northern Region is characterised by the low percentage of ethnic minorities, about 2% , and itsDuring the 50 years following the end of the Second World War, the economy of Newcastle transformed from one dominated by heavy industry and coal production and trade to a service based economy by the early 1990s. This paralleled remarkable changes in societal factors; for example, the 1967 Abortion Act, the National Health Service (Family Planning) Act (1967), availability of free family planning services irrespective of age or marital status from April 1974, the Sex Discrimination Act (1975) and the Employment Protection Act (1975) were introduced during the study period.The PAMPER database contains birth details on all singletons born during 1961–92 to mothers resident within the city of Newcastle upon Tyne in Northern England. Information on multiple births was also collected, however it was excluded from these analyses as multiplicity is a known risk factor for the outcomes of interest of the PAMPER study, i.e. preterm birth and low birthweight. The boundaries of the PAMPER study area are shown in Figure To capture home births, we additionally abstracted data from 'birth ledgers' 1961–1973), containing limited information on all births , the address at birth was assigned a postcode from the 1991 postcode book or a grid reference. This allowed us to locate enumeration district (ED) of mothers' place of residence and hence to obtain the Townsend Deprivation Score (TDS), an area-based measure of material deprivation , at ED lWe linked the dataset to information on stillbirths and infant deaths (including causes of death) from the Office for National Statistics (ONS) and to death data from the Northern Perinatal Mortality Survey (PMS) (available from 1981 onwards) . MultiplAs Gosforth in the north and some western residential parts of the PAMPER study areas were not part of the city of Newcastle upon Tyne prior to 1974, the ONS could not provide us with all stillbirths and infant deaths for these areas for this earlier period. However, we obtained death certificates and causes of stillbirth for cases known to us to be stillbirths and infant deaths. This may still have resulted in some missing infant deaths if a postneonatal death was not recorded in the hospital notes.Data entry staff  were trained in the medical terms/abbreviations used in the neonatal records and thus the percentage of errors was minimised. SVG was responsible for completing a descriptive 'summary' field, which contained the medical diagnosis and causes of death. In addition to the ONS and PMS data, stillbirth and infant death data were validated using birth record sources mentioned above.At the initial stage of data entry, we double entered approximately 1% of the estimated total of 120,000 birth records for different decades of the study period  to assess accuracy of the data entry results. At the final stage of data entry, the data were validated by checking for implausible values .Table For data capture we used the 4D database software suitable for a simultaneous data entry by several people, for data manipulation we used Microsoft Office Access 2003.Stillbirths included were all babies born dead at 28 or more completed weeks of gestation. There were 12 cases (1%) recorded as stillbirths by the ONS with uncertain gestational age which were also included. Stillbirths with birthweight less than 500 g were excluded if gestational age was unknown. Infant death was defined as a death, following live birth, of an infant under one year of age. We defined preterm birth as birth at a gestational age less than 37 completed weeks and term birth as birth at a gestational age ≥ 37 weeks.For descriptive statistical analysis we used the statistical software package SPSS for Windows, version 14.0. We used chi-square tests to test differences in proportions and independent-sample t-tests for comparison of means.The study received a favourable ethical opinion from the Sunderland Local Research Ethics Committee (SLREC 1071).The number of births was highest in the early 1960s, followed by a steady decline until the mid 1970s and a further increase in the 1980s Figure . Home biFigure There was a dramatic decline in both stillbirth and infant mortality over the three decades Table .Between 1961 and 1992 the average family size decreased, mainly due to a decline in the proportion of families with ≥ 3 children ] compared to 1971–80 [2308 g (SD ± 683)] in contrast to the first decade .Mean birthweight was lowest in the early 1960s, averaging around 3267 g in the second decade, followed by a gradual increase during the second half of the study period Figure . Thus duThe proportion of preterm births declined from 7% in 1961–70 to 6% in 1971–80 Figure , but it Table Our study using population-based birth data in a single conurbation over three decades reported that between 1961 and 1992, when stillbirth and infant mortality rates declined dramatically, maternal age, parity, birthweight and gestational age changed substantially.National trends on the total fertility rates for 1960–1990 mirror temporal trends shown in our study, where we used parity as a measure of fertility; during the 1960s 'baby boom', the national total fertility rates peaked in 1964 followed by a subsequent decline with a lowest level in the mid 1970s and a slight increase afterwards . It has We report a steady increase in the overall mean birthweight starting from the mid 1970s, which we observed for term births only and despite the increase in the proportion of preterm births in the second half of the study period. Hence, the observed rise in the total mean birthweight is likely to reflect the increase in birthweight for gestational age for term infants. This was also observed in Norway, where an increasing trend was reported for term births for 1967–1998 , but notThe proportion of preterm births declined in the second decade compared to the first, but it was followed by a steeper increase in 1981–92. To our knowledge, there are no population-based studies from the UK for comparison. Studies from other countries also reported the increase in the percentage of preterm birth from the 1980s ,36,37. STownsend deprivations scores, which we calculated for each birth in the database to measure neighbourhood socioeconomic status, also changed over time: the scores seemed to improve for the most affluent quintile and deteriorate for the most deprived, thus making the gap between the affluent and deprived groups wider. This is in line with the widening socio-economic and health inequalities which are now well documented in the UK.The population-based PAMPER birth record database contains historical high-quality birth data in a defined compact geographical setting over a 32 year period during which there have been significant changes in obstetric and neonatal services. The completeness of the PAMPER database both for the number of births and information collected for each birth is a major strength. National and local trends in the number of births in the UK confirm the temporal fluctuations also observed in the PAMPER study: the highest number of births at the beginning of the 1960s , followed by a decline in the 1970s and a further increase in the number of births during the 1980s . The comThe availability of accurate population-based gestational age, a major determinant of birthweight, is one of the leading strengths of the PAMPER database, as gestational age was not available in national birth statistics during the study period. Further, birthweight for live births was not collected in the UK at national level until 1975 . Without gestational age, interpretation of trends in birthweight could be misleading, as it is not possible to disentangle whether changes in birthweight are attributable to changes in rates of preterm birth or to changes in actual fetal growth. However, in the UK and elsewhere in the world there is a lack of information on the incidence of premature birth using accurate data by gestation .The accuracy of the data for the key variables was ensured by multiple checking, internal (within the database) and external  validation of the data.The PAMPER database also has several limitations. The lack of information on some important determinants of fetal weight at birth such as maternal height, maternal smoking and exposure to environmental tobacco smoke, which have changed over time thereby affecting changes in birthweight, is disappointing. For example, an increasing trend in maternal height was reported in Scotland for 1980–2000 . In the The accuracy of gestational age estimates is important for epidemiologic studies of pregnancy outcomes. Different methods for gestational age assessment  or early ultrasound measurements) throughout the study period may introduce bias in gestational age estimation over time. Thus it has been suggested that higher rates of preterm birth may be reported if determination of gestational age is based on ultrasonographic dating alone ,46. In tThis historical population-based study documents substantial temporal changes in key maternal and fetal factors affecting birth outcomes over a 32-year period during which much social change has taken place. The availability of accurate gestational age is extremely important for correct interpretation of trends in birthweight.The authors declare that they have no competing interests.SG carried out the statistical analysis and drafted the paper. All authors were co-investigators on the Wellcome Trust grant, contributed to the initiation of the project and study design, and commented on the drafts of the paper. All authors have read and approved the final version of the manuscript.The pre-publication history for this paper can be accessed here:


During the past two decades, the hypothesis of fetal origins of adult disease has received considerable attention. However, critique has also been raised regarding the failure to take the explanatory role of accumulation of other exposures into consideration, despite the wealth of evidence that social circumstances during the life course impact on health in adulthood. The aim of the present prospective cohort study was to examine the contributions of birth weight and life course exposures (cumulative socioeconomic disadvantage and adversity) to dyslipidemia and serum lipids in mid-adulthood.A cohort  was prospectively examined with respect to self-reported socioeconomic status as well as stressors  at the ages of 16, 21, 30 and 43 years; summarized in cumulative socioeconomic disadvantage and cumulative adversity. Information on birth weight was collected from birth records. Participants were assessed for serum lipids , apolipoproteins (A1 and B) and height and weight  at age 43. Current health behavior  was reported at age 43.Cumulative life course exposures were related to several outcomes; mainly explained by cumulative socioeconomic disadvantage in the total sample  and also by cumulative adversity in women . Birth weight was related only to triglycerides in women, independently of life course exposures, health behaviors and BMI. No significant association of either exposure was observed in men.Social circumstances during the life course seem to be of greater importance than birth weight for dyslipidemia and serum lipid levels in adulthood. Since the early 1990 s, the research field of life course epidemiology has studied the health effects of biological and social exposures during different periods of life and the corresponding long-term processes spanning over the life course . Low birEarly studies on the fetal origins of serum lipids and dyslipidemia indicated that lower birth weight was linked to an atherogenic lipid profile in adulthood -12. HoweIn conclusion, birth weight might only be weakly related to adult serum lipids. This has been interpreted as an argument for lifestyle during adulthood as a major determinant of atherogenic serum lipids . AlternaThe sample was based on the Northern Swedish Cohort, a 27-year prospective cohort study comprising all pupils in the ninth grade of the Swedish compulsory school living in Luleå in 1981, when the participants where 16 years of age . Follow-up surveys were conducted when the participants were 18 (1983), 21 (1986), 30 (1995) and 43 (2008) years of age. In this report, data from the 1981, 1986, 1995 and 2008 surveys are presented. The cohort has in various comparisons been found to be representative of the Swedish population . Of the Participants completed a comprehensive questionnaire at all follow-ups. Although the composition of the questionnaire varied at different ages due to the age-specific relevance of some topics, the main areas covered in all versions included health and health behaviors, social and socioeconomic conditions, school/working conditions and leisure activities. The majority of items originated from the Swedish Survey of Living Conditions  and the Delivery records of the participants' mothers from 1965  were retrieved from the archives of the respective delivery ward. From the delivery records, information on birth weight, length, date of delivery, date for last menstruation prior to delivery, and diagnoses according to the Swedish adaptation of the 6th revision of the International Classification of Disease , was doc3, kg/m3) were initially used as the main continuous measures of body size at birth. Since both ponderal index and birth length were less consistently related to lipid levels than was BW in preliminary analyses, these alternative measures were excluded from further analyses. Preterm or premature birth  did not influence the estimated associations, why all cases were included in the analyses. The associations between birth weight and lipids appeared to be approximately rectilinear by graphical inspection.Birth weight (BW) (kg), Birth length (cm), and ponderal index (weight/lengthcumulative adversity. The following components of adversity were included (k = 28):Adversity was defined as potentially stressful environmental exposures. Items representing this concept were selected from the questionnaires at each age, while items substantially dependent on subjective appraisal of the exposure were not selected. Since different versions of the questionnaire were used at different ages, the set of adversity components also varied between measurements. All non-binary components were dichotomized as close to the 80th percentile as possible. All components were then added generating an index of Parental illness (age 16) was defined as one or both parents having a physical illness, mental problems, and/or alcohol or drug problems, as reported by the adolescent.Parental loss (age 16) was defined as either having experienced parental separation/divorce (n = 181), or parents never living together (n = 17), or death of either parent (n = 28).Parental unemployment (age 16) was defined as one or both parents being unemployed or granted a disability pension at the time of the survey (housewives were classified as employed).Residential crowding (age 16) was defined as the participant not having her/his own room.Residential mobility (age 16 and 21). The participants were asked how many times they had moved in their lifetime (at age 16) and during the last three years (at age 21). High residential mobility was defined as > 2 moves (= 1), compared to 0-2 moves (= 0) at each time period.Unemployment  was defined as current own unemployment or disability pension, based on a question about current labor market situation.Spousal unemployment (age 30 and 43). The participants were asked if their partner had been unemployed during the last five years (age 30), and about the partner's current labor market situation (age 43). In the latter case, unemployment or disability pension was considered unemployment.Low cash margin . The participants were asked if it would be possible for them to raise a certain amount of cash within a week [n a week . The amoLow income (age 21) was defined as a self-reported monthly income <3,400SEK.Financial strain (age 30 and 43). The participants were asked how often they were forced, due to financial reasons, to abstain from any out of eleven different material needs . The response options were 'often', 'seldom', 'never' or 'not applicable'. The number of 'often' responses was added up to an index and dichotomized at the 80th percentile, resulting in 3 or more frequent cut-backs considered as high financial strain at age 30, and one or more at age 43.Illness and death of close ones . The participants were asked if someone close had had a serious or long-term illness, and if someone close had died, during the last three years (age 21), 12 months (age 30), or five years (age 43). A positive response was considered an adversity, separately for illness and death.Separation (age 30 and 43) was defined as having broken up from a long-term relationship involving co-habituation during the last 12 months (at age 30) or since the age of 30 (at age 43).Bullying or sexual harassment at work (age 30 and 43). The participants were asked how often they had been exposed to personal persecution at the workplace through mean words and actions from bosses or colleagues, or exposed to sexual harassment through unwelcome or degrading sexual insinuations. Due to the low frequencies, a couple of days per months of either bullying or sexual harassment was considered exposure (= 1).Exposure to threat and violence (age 30 and 43). The participants were asked if they had been exposed to threats of violence that were so serious that she or he was scared, and if they had been exposed to physical violence, during the last 12 months. Due to low frequencies, a positive response on either violence or threat was considered exposure (= 1).Social isolation (age 30 and 43). The participants responded to four questions from the Availability of Social Integration (AVSI) scale [th percentile was regarded as exposure to social isolation.I) scale ,33, concLow decision latitude (age 30 and 43). Participants responded to 6 items about decision latitude from the Swedish Demand-Control Questionnaire (DCQ) [th percentile was considered exposed to low decision latitude.re (DCQ) , with reParticipants' own occupation at age 21, 30 and 43 were coded according to the socioeconomic classification system of Statistics Sweden . The claCumulative socioeconomic disadvantage was operationalized as the number of life course periods with low SES (range 0-4).2) was calculated. For those who had not completed weight and height measurement (n = 9), BMI calculations were based on self-reported height/weight. There was a high correspondence between measured and self-reported height, r(891) = .91.Self-reported health behavior at age 43 was examined as potential confounders: daily smoking (yes/no), daily snuff use (yes/no), alcohol consumption. Alcohol consumption  was estimated from questions on the frequency of alcoholic beverage consumption, and approximate volume of beverage per occasion  and dichotomized at 110 g/week for men and 80 g/week for women . At age At age 43, participants were invited to a health examination at their local health care center, including blood sampling after an overnight fast. Samples were assessed with respect to total cholesterol (TC), low-density lipoprotein cholesterol (LDL-C), high-density lipoprotein cholesterol (HDL-C), triglycerides (TG) Apolipoprotein (Apo) A1 and Apo B , according to the routine methods at the Department of Clinical Chemistry, Umeå University Hospital, Sweden. TG concentrations were logarithmically transformed by the natural logarithm prior to analysis to achieve a normal distribution. The other outcomes displayed approximately normal distributions.Dyslipidemia was defined according to the 2003 European Society of Hypertension-European Society of Cardiology guidelines [idelines : total cidelines   when all variables were entered into the analysis . See Table Those with missing lipid data (n = 115) did not differ significantly from those with available lipid data (n = 888) regarding birth weight , cumulative adversity (p = .16) or cumulative SES (p = .60). Those with dyslipidemia medication (n = 28) did not differ from those without medication regarding SES, adversity or birth weight , and where therefore included in the analyses.Since different life course models might be relevant to explain socioeconomic influences on health, we employed a formal method recently proposed by Mishra et al.  for comp2 with corresponding F test. A higher ΔR2 provided by the first step than in the second step would indicate that birth weight plays an important role for adult lipids while life course exposures are of less importance. The reverse situation would point to life course exposures being of larger importance. Standardized regression coefficients (β) are presented in addition to unstandardized regression coefficients (b) and standard errors (SE), to provide comparable estimates for the independent variables measured on different scales. A substantial attenuation of the birth weight coefficients with the adjustment for life course exposures could be an indication of the latter acting as confounders or mediators. Attenuation of life course coefficients in the third step would indicate that the association might be explained by behavioral pathways. Since adult BMI could act as a mediator or a confounder of birth weight [As our main analyses, associations between dyslipidemia and birth weight, cumulative socioeconomic disadvantage and adversity and current health behavior, were first examined by means of logistic regression analysis; results are presented as crude and adjusted Odds Ratios (OR) and 95% confidence intervals (CI). This general approach was then extended to specific examinations of the serum levels of each lipid. For these analyses, hierarchical linear regression was used, with continuous lipid measures at age 43 regressed on birth weight (Model 1), with the subsequent addition of cumulative socioeconomic disadvantage and adversity (Model 2), and current health behavior (Model 3). Additional contribution provided by each step was appraised by the ΔRh weight  as well h weight , in a coAnalyzes were conducted on the whole sample as well as on women and men separately. SPSS version 17.0 was used for all analyses.As can be seen in Table As an initial examination of linear relationships between the exposures and the lipid levels, bivariate correlations between the continuous variables of interest are shown in Table These bivariate analyses were extended to hierarchical linear regression analyses. Total cholesterol was not explained by the set of predictors to any important degree, in either women  or men ). Similarly, LDL-C was not related to any of the independent variables in women  or men . Summaries of the results for the remaining outcomes are found in table In the total sample table , birth wIn women Table , birth wIn men Table , associaIn complementary analyses, BMI at age 43 was added to the final model (Model 3) for each of the lipid outcomes (data not shown). BMI displayed a significant independent association with all lipids  in the total sample, and with all lipids (p < .001) except for total cholesterol (p > .10 in both women and men) and LDL-C in men (p > .10) in the sex-stratified analyses. The contribution of SES in the total sample was substantially attenuated by the addition of BMI, for HDL-C , triglycerides  Apo A1  and Apo B . In women the significant/borderline significant independent contributions of adversity were unaffected by the addition of BMI for all lipids, as was the contribution of birth weight for triglycerides . In men, the near-significant contributions of birth weight were slightly accentuated and reached significance for both HDL-C  and Apo A1 (β = .10 p = .041). There were no other substantial changes of coefficients.This study provides support for the hypothesis that the accumulation of unfavorable circumstances over the life course is related to adult dyslipidemia. Similar results were shown for serum levels of some lipids  but not others . Associations were mainly found among women. Support for the fetal origins hypothesis was only found for triglycerides in women. Thus, life course exposures provided a more substantial overall contribution to adult serum lipids than did birth weight.In the total sample, birth weight was not related to dyslipidemia or to any specific lipids, while cumulative socioeconomic disadvantage displayed significant associations with dyslipidemia, HDL-C, triglycerides and Apo B. For dyslipidemia, the independent contribution of SES and adversity were nonsignificant after mutual adjustment, possibly due their interrelationship, and the ORs were further attenuated by the addition of health behaviors. For HDL-C and triglycerides these associations seemed to be independent of current health behavior. These results support previous indications of weak and inconsistent influences of birth weight on lipids , and genThe separate analyses on women and men provided a partly different picture. In women, triglycerides were significantly related to birth weight, an association independent of both social and behavioral factors. This is in line with previous findings indicating triglycerides to be the lipids with the most consistent association to birth weight , althougAdult BMI strongly attenuated the contribution of SES in the total sample, consistent with BMI acting as a mediator in the SES-lipids relationship. In men, the contribution of birth weight was slightly accentuated after the addition of BMI in adulthood, a phenomenon that has been highlighted in the literature on association between birth weight and lipids  and blooTotal cholesterol and LDL cholesterol displayed no association with either birth weight or life course exposures. This is consistent with the observations that birth weight is only insubstantially related to total cholesterol levels . MoreoveThe main strengths of this study are the prospective design and high response rate. The high response rate, achieved despite the long-term prospective design, protects against selection bias which otherwise could influence the results . In addiThe simple dichotomization of socioeconomic status at each age could possibly mask variation of SES important for the studied outcomes. However, this approach is a conventional operationalization of the cumulative risk model ,52. AlthThe meager findings concerning birth weight could be a result of too small a sample size and insufficient power. However, the aim of the study was to examine the relative contributions of different classes of exposures, and despite mainly non-significant results regarding birth weight, both life course exposures and health behaviors displayed more consistent associations with the outcomes. Thus, the relatively lower explanatory value of birth weight compared to life course exposures can still be interpreted. As noted, these findings are in line with previous research concluding that although there might be an influence of birth weight on adult cardiovascular risk factors, it is dubious that this potential effect is of substantial practical importance in this area ,14.The accumulation of adverse social circumstances over the life course was found to be of relevance for dyslipidemia and serum levels of some lipids in adulthood, while the contributions of birth weight were inconsistent and where present, eclipsed by the stronger contribution of accumulated life course exposures. These results indicates that although there might be a weak influence of fetal conditions, social conditions and experiences over the life course do play a greater role for cardiovascular risk factors in adulthood.The authors declare that they have no competing interests.AH was responsible for the conception and design of the study and UJ and TT contributed to the design of the study. AH and UJ were responsible for the data collection. PEG was responsible for the statistical analyses and writing the first draft of the manuscript. All authors participated in interpreting the results, provided input on manuscript drafts, and read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2458/10/484/prepub


Choledochal cyst is an uncommon congenital disease of the biliary tract in the UK. There are five main types of choledochal cyst with several recognised sub-types. However, occasional variants do occur.We report a case of a female infant with an antenatally diagnosed choledochal cyst. The operative cholangiogram revealed an unusual intrahepatic biliary tree. The cyst was successfully excised and the infant is well at 18-months follow up.The anatomy should be clearly defined before surgical excision as abnormal variants can occur, which usually do not fit into the known classification types and subtypes. Choledochal cysts are congenital abnormalities of the biliary system and consist of cystic dilatations of the extrahepatic biliary tree, intrahepatic biliary ducts or both. The incidence of choledochal cyst in western countries has been reported to be 1:100,000 to 1:150,000 ,3. CholeA female Caucasian infant was born at term with an antenatal diagnosis of an abdominal cyst. Prenatal diagnosis could not confirm the exact origin of the cyst. A prenatal differential diagnosis of choledochal cyst, duplication cyst and omental cyst was given. Due to diagnostic difficulties, several post-natal scans were performed before the cyst was confirmed to be a choledochal cyst. The first two scans showed a cyst in the subhepatic area which was septated and loculated and was displacing the intrahepatic vessels. The common bile duct was found to be of normal calibre. There was no intrahepatic dilatation. It was reported to be a hepatic cyst or a choledochal cyst but with features not characteristic of either. She remained asymptomatic and once the diagnosis of a choledochal cyst was confirmed, an elective excision of the cyst was performed at the age of seven months. At laparotomy, a large 5 cm × 4 cm × 2 cm cyst was found at the level of the porta hepatis with normal hepatic and common bile duct Figure  which diHistopathology showed an inflamed cystic structure lined with stratified squamous epithelium and containing areas of probable metaplasia. The liver biopsy showed no evidence of biliary atresia.Choledochal cyst is the cystic dilatation of extrahepatic bile ducts, with or without the dilatation of the intrahepatic biliary tree. It may be associated with biliary atresia .Based upon the clinical and anatomic findings, in 1959, Alanso-Lej, Revor and Pessagno classified choledochal cysts into three main types . Type 1 Our patient is unusual in that the anatomy does not fit with any of the classical descriptions of choledochal cysts. The cholangiogram finding of a biliary blush was also unusual. The surgical management did not differ from a conventional choledochal cyst. However, surgeons should be aware of the possibility of variant anatomy in choledochal cysts, which is outside the normal classification for this condition.Written informed consent was obtained from the patient for publication of this case report and any accompanying images. A copy of the written consent is available for review by the Editor-in-Chief of this journal.The authors declare that they have no competing interests.KL was responsible for the study cocept & design & for final approval of the version for publication. BN was involved in patient care & helped in review of literature. JS was responsible for acquiring the data, review of literature and drafting & revising the manuscript.


We examined the differential associations of each parent's height and BMI with fetal growth, and examined the pattern of the associations through gestation. Data are from 557 term pregnancies in the Pune Maternal Nutrition Study. Size and conditional growth outcomes from 17 to 29 weeks to birth were derived from ultrasound and birth measures of head circumference, abdominal circumference, femur length and placental volume (at 17 weeks only). Parental height was positively associated with fetal head circumference and femur length. The associations with paternal height were detectible earlier in gestation (17–29 weeks) compared to the associations with maternal height. Fetuses of mothers with a higher BMI had a smaller mean head circumference at 17 weeks, but caught up to have larger head circumference at birth. Maternal but not paternal BMI, and paternal but not maternal height, were positively associated with placental volume. The opposing associations of placenta and fetal head growth with maternal BMI at 17 weeks could indicate prioritisation of early placental development, possibly as a strategy to facilitate growth in late gestation. This study has highlighted how the pattern of parental–fetal associations varies over gestation. Further follow-up will determine whether and how these variations in fetal/placental development relate to health in later life. The latter is reflected onto the fetus' own nutritional status in what we refer to as intrauterine effects. There is evidence that parental anthropometry is strongly associated with birth size. For example, maternal BMI and height explain a large proportion of the geographical variation in birth weight, length and head circumference A limitation of birth anthropometry as a proxy for fetal growth is that two babies of a similar size at birth may have had radically different growth trajectories in-utero a posteriori analysis, we also examined the relationship with placental volume at 17 weeks gestation.Data from mother–father–fetus/offspring trios from the Pune Maternal Nutrition study were used to: [1] examine the association between parental height and BMI and fetal head circumference, abdominal circumference and femur length, [2] assess the relative influence from the mother and father, and [3] explore the timing and behavior of these associations over gestation. In an 22.1Data are from the Pune Maternal Nutrition Study (PMNS), a rural Indian community-based prospective study. The cohort was established from a house-to house survey of all married women of childbearing age (15–40 years) living in 6 villages located 40–50 km from Pune City (n = 2675). Enrolment took place between 1994 and 1996; 2466 women (92%) agreed to participate and 1102 became pregnant. The majority of women were vegetarian and had below recommended intakes of energy and protein. Further details are elsewhere 2.22.2.12.Each woman's height and weight were measured by health workers every three months until pregnancy occurred; the last set of measurements prior to conception was used as pre-pregnant values. Paternal height and weight were also measured within 3 months of the confirmation of the woman's pregnancy. BMI was calculated as wt/htSeveral potential confounding variables were examined. Information on household socio-economic status was collected using a standardised questionnaire 2.2.2Each woman was visited monthly by a trained health worker to record the date of the last menstrual period (LMP). Women who reported a missed period underwent an ultrasound examination 15–18 weeks after their LMP to confirm pregnancy. A further ultrasound scan was scheduled for 28 ± 2 weeks gestation. The median gestational age at each examination was 17.1  and 29.4  weeks.On each visit, one of two trained sonologists (MCC and ASK) obtained measures of fetal head circumference (HC), biparietal diameter (BPD), abdominal circumference (AC) and femur length (FL). Scans were carried out using a portable machine fitted with a curvilinear array 5 MHz transducer . Placental volume was measured at the first ultrasound scan only, with a linear-array 3.5 MHz transducer with a foot length of 14 cm and a 12.5 cm field of vision, using a modified planimetric technique Health workers performed detailed anthropometry of the babies within 72 hours of birth using standardised techniques. Birthweight was measured using a Salter spring balance; crown–heel length was measured using a portable Pedobaby Babymeter ; occipito-frontal head circumference and abdominal circumference were measured using a fibre glass tape , the latter immediately above the umbilical cord insertion, in expiration.2.3For the purpose of this analysis, we used LMP dates to estimate gestation — this is important in a study of fetal growth, because ultrasound dating assigns a gestational age based on fetal size, thus erasing variation in growth during early pregnancy. However, to minimize errors from erroneous LMP dates, fetuses whose gestational age estimated by ultrasound differed by more than 2 weeks from the LMP estimate were excluded (n = 144). Sonographic gestational age was determined on the first visit, using a prediction equation based on fetal BPD, AC and FL 2.4From 1102 pregnancies, in addition to those excluded due to gestational dating discrepancies, 288 were excluded due to spontaneous abortions, fetal anomalies on ultrasound, multiple pregnancies, medical terminations or pregnancies detected later than 20 weeks gestation . Babies 2.5Exposures and outcomes were standardised to a z-score so that we could compare the relative associations from the mother and father, and across fetal components. Parental BMI was log transformed before standardisation. Growth models were developed to describe the relationship between the fetal variables  and gestational age using the method described by Royston (1995) 2.62.6.1Our analysis strategy was to assess associations of parental height and BMI with fetal size at scan 1 (17 weeks), then determine the association with growth from scan 1 to scan 2 (17–29 weeks) independent of size at 17 weeks, and move forward again to assess associations with growth from 29 weeks to birth independent of size at 17 and 29 weeks. To do this, we created 2 conditional growth variables for each fetal component — size at 29 weeks conditional on size at 17 weeks (29| 17 wks), and size at birth conditional on size at 29 and 17 weeks (Birth| 29 and 17 wks), by regressing each size measurement (z-score) on earlier size (stratified by sex) and keeping the standardised residuals. The conditional z-score is a measure of growth velocity between 2 time points (e.g. 17–29 weeks) on a theoretical distribution which compares each fetus against other cohort members of the same size at time point 1 (e.g. 17 weeks). It can be interpreted as growth above or below that expected given earlier size. The set of variables  thus contain information on whether a fetus grew quickly or slowly for the intervals 0–17, 17–29 and 29-birth relative to other fetuses of the same growth history. Importantly, by construction, the set of conditional variables are uncorrelated, and so any association between an exposure and an interval is independent of its association with other intervals.2.6.2The outcome variables were the growth set for HC, AC, FL and placental volume at 17 weeks. We had no data on FL or leg length at birth so used birth length as a proxy for birth femur length; these have been shown to be closely related 2.6.3False paternity will dilute the paternal offspring relationship and bias the comparisons of the mother–offspring and father–offspring relationship in favour of showing larger maternal effects. We performed a sensitivity analysis to examine the effect of non-paternity using a previously described method 32, WHO guidelines). Most were younger than 22 years and one-third were primiparous. Fathers were also short and thin  was associated with a 0.11 SD increase in the part of HC growth from 17 to 29 weeks not associated with HC size at 17 weeks. There was no evidence for an additional effect of paternal height from 29 weeks beyond that which occurred earlier, however, paternal height was related to unconditional HC size at birth .There was no evidence of an association between maternal height and any AC growth or size parameters. There was a very weak suggestion of a positive association between paternal height and AC at 17 weeks. Paternal height was not related to AC in any of the other independent growth intervals or to AC size at 29 weeks or birth .There were no associations between maternal or paternal height and femur length (FL) at 17 weeks, or conditional growth from 17 to 29 weeks . Both maDespite a suggestion that the associations with paternal height were apparent earlier in gestation, there was no evidence for a difference between maternal–offspring and paternal–offspring relationships for any period of gestation or for any fetal component .3.2Pre-pregnancy maternal BMI was negatively associated with HC size at 17 weeks (β (per SD) = − 0.11 SD; 95% CI: − 0.19, – 0.02) and positively associated with conditional HC growth from 29 weeks to birth (β (per SD) = − 0.09 SD; 95% CI: 0.0, 0.18) . There wPaternal BMI did not appear to be related to any fetal growth or size component , and des3.3The median (IQR) placental volume at 17 weeks gestation was 148.6 ml . Maternal BMI ((β (per SD increase) = 0.14 SD; p = 0.003) and paternal height ((β (per SD increase) = 0.1 SD; p = 0.02) were associated with placental volume at 17 weeks. There was no evidence for a relationship with maternal height or paternal BMI . Placent3.4Increasing the rate of non-paternity drew the putative father's association towards the mother's. For example, with non-paternity assumed at 15% the association between paternal height and growth in length from 29 weeks to birth changed from 0.13SD (per SD increase in father's height) to 0.16SD, the mother's coefficient reduced by only 0.004SD from 0.23SD. However, the effect of non-paternity was very small when the putative father's association was close to zero. Thus the patterns where there is a suggestion of an association for the mother but no clear evidence for the father, for example the associations with placental size, were unaffected by rates of non-paternity.4To the best of our knowledge, this is the first study to examine parental height and BMI associations with ultrasound measures of fetal growth in a rural Indian population with a high prevalence of low birth weight. Our results suggest that associations between paternal height and fetal head circumference and femur length are detectible earlier in gestation (17–29 weeks) compared to the same associations with maternal height. Fetuses of mothers with a higher BMI had smaller head circumferences at 17 weeks, but head growth caught up to become larger in late gestation. Paternal BMI showed no significant associations with any component of fetal growth. Maternal but not paternal BMI, was positively related to placental size at 17 weeks.Our study has several important strengths. First, we used serial fetal biometry to characterise prenatal growth as opposed to birth weight. Second, ultrasound measurements were made by only 2 sonographers and intra and inter-observer reliability was excellent. Third, we were able to obtain accurate LMP dates since health workers visited women every month prior to pregnancy. Fourth, the maternal BMI data are likely to be an accurate representation of pre-pregnant BMI as data were collected within a 3 month window prior to pregnancy. Fifth, a sensitivity analysis showed that non-paternity was unlikely to have affected our main findings. And last, the use of conditional growth variables as a model of fetal growth allowed an exploration of the onset and pattern of parent–fetal associations.The main limitation is a lack of power, a bigger study may have exposed some of the trends in the timing of associations over gestation that we reported, however intergenerational data with fetal ultrasound from populations in rural India are rare. We were also restricted to investigating growth intervals at relatively fixed time points. However, while more densely spaced measures may have offered a deeper insight, it does increase signal-noise ratio in the growth variable.Maternal height was positively related to conditional growth in the last gestational interval (29 weeks — birth) for head circumference and femur/birth length. A study of American singletons also found positive associations between maternal height and head circumference at 31 weeks, and with femur length at 25 weeks Positive correlations of maternal and paternal height with newborn skeletal measurements like length and head circumference are well described For maternal BMI, there was a positive association with HC growth in the 29 week to birth interval, consistent with findings from Goldenberg et al. (1993) The different patterning of associations reflected by the mother and father's BMI on placental volume and the suggestion of a different pattern of associations on the fetal components — particularly head circumference, suggest some direct and potentially modifiable effects of maternal nutritional status on placental and fetal growth. Higher maternal BMI may promote placental growth in early-mid pregnancy, and this may enhance fetal growth in late pregnancy, although this interpretation of our findings is speculative. We do not yet know the implications of variations in placental and fetal growth for future health in the child. Recent studies in the Helsinki birth cohort suggest that placental size and shape at birth, newborn size, and ratios of placental to newborn size, predict adult hypertension, and that these associations are conditioned by maternal nutritional status We recommend that epidemiological studies exploit the use of ultrasound to characterise variations in fetal and placental growth, and their timing during gestation. Such measures will allow more carefully designed analyses into questions related to the developmental origins of disease, and offer insight beyond that possible with birth weight alone.


Audit of disease and mortality patterns provides essential information for health budgeting and planning, as well as a benchmark for comparison. Neonatal mortality accounts for about 1/3 of deaths < 5 years of age and very low birth weight (VLBW) mortality for approximately 1/3 of neonatal mortality. Intervention programs must be based on reliable statistics applicable to the local setting; First World data cannot be used in a Third World setting. Many neonatal units participate in the Vermont Oxford Network (VON); limited resources prevent a significant number of large neonatal units from developing countries taking part, hence data from such units is lacking. The purpose of this study was to provide reliable, recent statistics relevant to a developing African country, useful for guiding neonatal interventions in that setting.This was a retrospective chart review of 474 VLBW infants admitted within 24 hours of birth, between 1 July 2006 and 30 June 2007, to the neonatal unit of Charlotte Maxeke Johannesburg Academic Hospital (CMJAH) in Johannesburg, South Africa. Binary outcome logistic regression on individual variables and multiple logistic regression was done to identify those factors determining survival.Overall survival was 70.5%. Survival of infants below 1001 grams birth weight was 34.9% compared to 85.8% for those between 1001 and 1500 grams at birth. The main determinant of survival was birth weight with an adjusted survival odds ratio of 23.44 (95% CI: 11.22 - 49.00) for babies weighing between 1001 and 1500 grams compared to those weighing below 1001 grams. Other predictors of survival were gender , birth before arrival at the hospital (BBA) , necrotising enterocolitis (NEC) , hypotension  and nasal continuous positive airways pressure (NCPAP) .Survival rates compare favourably with other developing countries, but can be improved; especially in infants < 1001 grams birth weight. Resources need to be allocated to preventing the birth of VLBW babies outside hospital, early neonatal resuscitation, provision of NCPAP and prevention of NEC. Neonatal mortality accounts for 37% of deaths below the age of 5 years and "improved neonatal and maternal care could save the lives of countless newborns" [e births . The sure births  to more e births , which cThere has been steady improvement in the overall early survival of VLBW infants over time  The purpose of this study was to review the survival to hospital discharge and morbidity of VLBW infants at Charlotte Maxeke Johannesburg Academic Hospital (CMJAH), a busy neonatal unit in a developing country.This was a retrospective record review of all neonates with a birth weight < 1501 grams admitted to the neonatal unit of CMJAH within 24 hours of birth from 1 July 2006 to 31 June 2007. All inborn neonates were admitted directly to a labour ward nursery, so statistics included inborn babies who died shortly after birth. VLBW babies who were delivered at outlying primary level hospitals or clinics and those who were born before arrival in hospital (BBA) were also admitted to the neonatal unit. Data was entered from hospital records onto a Microsoft Access (2003) database. Maternal information obtained from the delivery records included age, parity, gravidity, antenatal care, administration of antenatal steroids, syphilis screening and treatment, human immunodeficiency virus (HIV) screening and prophylaxis, place of delivery, fetal presentation and mode of delivery. HIV screening followed a protocol of voluntary counseling and testing; mothers could refuse to be tested. Prophylaxis was only given to infants where mothers were proven to be HIV positive. Polymerase chain reaction (PCR) testing to confirm HIV infection in the neonate was only done from 6 weeks of chronological age. Neonatal intensive care unit (NICU) admission was not determined by HIV exposure.The baby's weight, Apgar scores and details of delivery room resuscitation were also obtained from the delivery records. Gestational age was determined from a combination of maternal history  and the Ballard score, which was done by attending clinical staff. The birth weight was plotted on Fenton  growth chttp://www.ppip.co.za. The PPIP was established in 1999 in South Africa as a national tool for perinatal death audit. In order to have manageable data, the single most likely cause of death is listed - major categories include prematurity, asphyxia, infection and congenital anomaly. Each category is further subdivided into sub-categories; prematurity is subdivided into extreme immaturity, HMD, IVH, NEC and pulmonary haemorrhage. No postmortem examinations were done on the study patients. Details of ICU admissions include indication for ventilation, dates and type of ventilatory support, (IPPV or NCPAP) and surfactant therapy. Babies who received both NCPAP and IPPV were classified as needing ventilator assistance for the purposes of analysis.The cause of death was reviewed by the primary author (DEB) and classified according to the PPIP classification Babies were managed according to the unit policies at the time. Ventilatory support was offered to babies above 900 grams birth weight, due to severely limited tertiary resources. Babies were not routinely intubated or given NCPAP in the delivery room; ventilatory support (including NCPAP) was commenced when the infant showed signs of respiratory failure. All babies, irrespective of birth weight, were provided with standard neonatal care . Surfactant therapy was only given as rescue therapy to babies on ventilatory support, usually to those patients who did not wean rapidly from supplementary oxygen. A second dose of surfactant could be given if the baby had responded to the initial dose and then deteriorated again. NCPAP was introduced to the neonatal unit March 2006. During the period of the study, there was no rooming in facility, so mothers could only do KMC intermittently during the day. KMC was introduced once a baby was in room air and tolerating full enteral feeds. Cranial ultrasound was done during the first week of life by a paediatric neurologist and, if indicated, repeated after 1 to 2 weeks and just prior to discharge. Babies who died within the first 72 hours may not have undergone a cranial ultrasound. Screening for retinopathy of prematurity was done by an ophthalmologist at 36 weeks post conceptional age. If babies were discharged prior to this age, an outpatient appointment was booked for the ophthalmology clinic. Babies were discharged home once they had established enteral feeds, were off supplemental oxygen, maintaining temperature and had achieved a weight of 1600 grams. Some babies were discharged to regional step down facilities for weight gain, close to the time of discharge home.http://www.spss.com). Continuous variables were summarised using mean and 95% confidence intervals, while categorical variables were summarised as ratios and percentages. For the purposes of analysis, babies transferred out and those discharged home directly were combined as "survivors" and compared to those babies that died during their hospital admission. Cross-tabulations of categorical variables with survival were produced and statistical associations between these categorical variables and survival outcome were done using the Chi-Square test of association. Normally distributed continuous variables were compared using the unpaired t test and the Mann-Whitney U test was used to compare discrete variables and those continuous variables that were not normally distributed. Binary outcome logistic regression was done on individual variables to predict survival. Those variables which were significant at the univariate analysis were entered into a multiple logistic regression using the backward selection procedure. All the statistical tests were conducted at 5% significance level.Statistical analysis was done on a personal computer using SPSS version 17  of VLBW babies born to 448 mothers were retrieved and available for review.The overall survival was 334/474 (70.5%). The mean birth weight was 1133.5 grams (95% CI 1111.9 - 1155 .0), mean gestational age was 29.9 weeks (95% CI 29.6 - 30.1) and mean duration of hospitalisation was 25.8 days (95% CI 23.8 - 27.8). The mean age at time of death was 5.77 days (95% CI 3.66 - 7.88) and of discharge/transfer was 34.23 days (95% CI 32.13 - 36.32). The mean duration of supplemental oxygen was 8.2 days (95% CI 6.8 - 9.7) and mean duration of mechanical ventilation was 8.08 days (95% CI 6.15 - 10.01).The mean birth weight of survivors  was significantly greater (p < 0.001) than that for babies that died . The mean gestation period for survivors  was significantly more advanced than that of those who died . The median 5 minute Apgar score of those babies that survived, 8 (IQR: 1-9) was significantly higher than the non survivors, 6 (IQR: 1-10) with p = 0.005. Mortality by birth weight category is shown in figure Survival was closely related to birth weight category - ranging from zero below 601 grams to 62% (32/52) from 901 to 1000 grams and 93% (67/72) from 1301 to 1500 grams. The survival of extremely low birth weight infants (< 1001 grams) was 34.9% (50/143) compared to 85.8% (284/331) for babies with a birth weight from 1001 to 1500 grams. The adjusted survival odds ratio was 23.44 (95% CI: 11.22 - 49.00) for babies weighing from 1001 to 1500 grams compared to those weighing below 1001 grams.The main cause of death according to the PIPP classification is shown in Table The mean maternal age was 26.5 years (95% CI: 25.8 - 27.1) and 37.3% (152/407) were primiparous. Risk factors for mortality related to antenatal care, labour and delivery are presented in Table Odds ratios with 95% confidence intervals are presented for each risk factor. Significant predictors of survival on univariate analysis were size for gestational age, gender, antenatal care, place of delivery, mode of delivery, the need for delivery room resuscitation and hypothermia at birth. Maternal HIV exposure, maternal infection with syphilis, the administration of antenatal steroids and the presenting part at delivery did not predict survival. Only 69% of mothers had known HIV status and 36.5% of mothers had received antenatal steroids.Air leak was recorded in 3 patients (0.6%) and high frequency ventilation was used in 3 patients (0.6%). These variables were not included in the analysis due to small numbers. Bronchopulmonary dysplasia (BPD) (defined as oxygen requirement > 28 days) was present in 42 (8.8%) of babies. Nevirapine prophylaxis was given to 97/120 (80.8%) of the HIV exposed infants; 25/354 (7%) of mothers offered HIV testing refused consent. 11/474 (2.3%) of the babies were put up for adoption. Only 87 (18.3%) of the babies were screened for ROP prior to discharge-only two babies had ROP, both of which were stage 1. KMC was done by 211/474 (44.5%) of mothers.Odds ratios and 95% confidence intervals for each risk factor related to the neonatal period (both disease and treatment) are shown in Table Multivariate logistic regression is shown in Table This retrospective review provides current survival rates and indicates where resources should be channeled in order to improve survival of VLBW infants in South Africa. The overall survival rate was 70.5% for VLBW infants at CMJAH 2006/2007. This is almost exactly the same as that reported from CH Baragwanath for 2000 - 2002 71%),  which re%,  whichThe main determinants of survival in the present study - birth weight, gender, being born before arrival in the hospital, resuscitation at birth, NEC, hypotension and NCPAP are not surprising and very similar to other reports on VLBW outcome -7,9-22. South Africa is a developing country with limited health resources and high patient numbers; it is not possible to provide full tertiary support to every VLBW infant. For many years this problem has been addressed by limiting ventilatory support, including the administration of surfactant and NCPAP, to those neonates above a specified birth weight cut off. Prior to the widespread use of NCPAP, this cut off was 1000 grams. The ventilation cut off was reduced to 900 grams just before the study period, with the introduction of NCPAP to the unit. The poor survival or our ELBW infants is undoubtedly influenced by this policy and NCPAP with surfactant should be provided to babies from 750 grams in order to bring our ELBW survival rate up to that in other developing countries ,20,21.NEC and hypotension were the other significant predictors of mortality. The overall rate of NEC was 5.5%, which accounted for 10% of the deaths. Interestingly the rate of NEC was comparable to that in the VON in 2005, which represents well resourced settings. Prevention of NEC should also be a priority, including promotion of breastfeeding. Our rate of breastfeeding at the time of the study was extremely low, mainly due to the HIV epidemic. Only 69% of mothers were tested for HIV, but 36% were positive. The protocol during the study period was to formula feed HIV exposed babies. Of concern, is the high rate of untested mothers, the relatively high rate of refusal to be tested and failure to administer HIV prophylaxis in 16.4% of exposed mothers; this may reflect the mothers who were diagnosed after the early neonatal period. Ensuring that all mothers are counseled and tested for HIV is essential to prevent mother to child transmission and, in turn, facilitate the promotion of breastfeeding and reduction of NEC. Although 36% of screened mothers were HIV positive, HIV status did not predict neonatal outcome. This is in agreement with a study from Durban  which foCMJAH represents a high risk obstetric population with many referrals for obstetric complications, the most frequent of which is pregnancy induced hypertension. Improved management of this obstetric complication may reduce the number of VLBW infants. There are also significant social problems with 2.6% of the babies delivered as a result of illegal termination of pregnancy and 2.3% being given up for adoption. The whole issue of preventing unwanted pregnancy is also pertinent in our population.The low rate of in hospital screening for ROP is also an area of concern and needs to be improved. A significant number of babies are referred to the ophthalmologist at the time of their first follow up visit for ROP screening, but this may be too late in terms of adequate intervention.Although the overall survival of VLBW infants in our unit compares with the global average  where the emphasis is on service delivery.AGA: appropriate for gestational age; BBA: born before arrival at the hospital; BPD: bronchopulmonary dysplasia; CH: Chris Hani; CMJAH: Charlotte Maxeke Johannesburg Academic Hospital; CS: Caesarean section; ELBW: extremely low birth weight infants (< 1001 grams); HIV: human immunodeficiency virus; HMD: hyaline membrane disease; IPPV: intermittent positive pressure ventilation; ICU: intensive care unit; IVH: intraventricular haemorrhage; LGA: large for gestational age; NEC: necrotising enterocolitis; NIHCD: National Institute of child health and development; NCPAP: nasal continuous positive airways pressure; PDA: patent ductus arteriousus; PVL: periventricular leukomalacia; ROP: retinopathy of prematurity; SGA: small for gestational age; VLBW: very low birth weight; VON: Vermont Oxford Network.The authors declare that they have no competing interests.Data collection and manuscript preparation was done by DEB; statistical analysis was done by DEB and TFC while PAC was involved in manuscript preparation. The final manuscript was approved by all authors.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2431/10/30/prepub


There is increasing evidence indicating that genes involved in certain metabolic processes of cardiovascular diseases may be of particular influence in people with low body weight at birth. We examined whether the apolipoprotein (APO) E, APOC3 and the peroxisome proliferator-activated receptor-γ-2 (PPARγ2) polymorphisms influence the association between low birth weight and blood lipid levels in healthy adolescents aged 13–18.5 years.A cross-sectional study of 502 Spanish adolescents born at term was conducted. Total (TC) and high density lipoprotein cholesterol (HDLc), triglycerides (TG), apolipoprotein (apo) A and B, and lipoprotein(a) [Lp(a)] were measured. Low density lipoprotein cholesterol (LDLc), TC-HDLc, TC/HDLc and apoB/apoA were calculated.Low birth weight was associated with higher levels of TC, LDLc, apoB, Lp(a), TC-HDLc, TC/HDLc and apoB/apoA in males with the APOE ε3ε4 genotype, whereas in females, it was associated with lower HDLc and higher TG levels. In males with the APOC3 S1/S2 genotype, low birth weight was associated with lower apoA and higher Lp(a), yet this association was not observed in females. There were no associations between low birth weight and blood lipids in any of the PPARγ2 genotypes.The results indicate that low birth weight has a deleterious influence on lipid profile particularly in adolescents with the APOE ε3/ε4 genotype. These findings suggest that intrauterine environment interact with the genetic background affecting the lipid profile in later life. Alimentación y Valoración del Estado Nutricional en Adolescentes, Food and Nutritional Status in Spanish Adolescents) indicated that birth weight is associated with a number of cardiovascular disease risk factors in adolescents such as total and central body fat, fat free mass, and muscular fitness  were measured. The coefficients of variation were less than 3% and the intra-class coefficients were higher than 0.96% for all blood variables. Quality control of the assays was assured by the Regional Health Authority, as is compulsory for all hospital clinical laboratories in Spain.Low density lipoprotein cholesterol (LDLc) was calculated with the Friedewald formula . The folBstU-I restriction enzymes as previously described [Genomic DNA was extracted and purified from 500 μL of whole blood treated with EDTA, using the Quiagen procedure described by Higuchi . APOE geescribed , and werescribed  . Further analyses were done after controlling for age, pubertal stage, socioeconomic status, gestational age, BMI and cardiorespiratory fitness (adjusted model). A separate regression model was performed for each lipid parameter. Additional analyses were also performed by genotypes.To stabilize variability and to achieve normality in the residuals, TC, HDLc, LDLc, TG, apoA, apoB and Lp(a) were transformed to the natural logarithm.1, P2, ..., Pi, ..., Pk. If P1 > α/k; none of the k tests are significant, and the test procedure is finished. If P1 ≤ α/k, test 1 is significant, and now P2 is examined. If P2 > α/(k - 1), none of the (k - 1) remaining tests are significant, but if P2 ≤ α/(k - 1), test 2 is significant and P3 is examined. This procedure goes on until Pi > α/(k - i + 1), and the procedure is interrupted. This method keeps family error rate less than α. Family error rate is defined as the probability that one or more false significances out of k tests is less than or equal to α.The impact on blood lipid levels of APOE, APOC3 and PPARγ2 genotypes was analyzed by one-way analysis of variance (ANOVA), for males and females separately. For the APOE genotypes, the subgroup means were compared by Tukey's test. Comparisons were adjusted for mass significance as described by Holm ,34. The Interaction effects between birth weight and sex, and between birth weight and genotypes were tested by inserting product terms for the relevant variables. The analyses were performed using the Statistical Package for Social Sciences  and the level of significance was set to 0.05.Descriptive characteristics of the study sample are shown in Table There was a significant  interaction effect between birth weight and sex in all the outcome variables, therefore, all the analyses were performed separately for males and females. There was a significant interaction effect between birth weight and APOE in all of the outcome variables , P = 0.051; TC-HDLc, P = 0.025; apoB/apoA, P = 0.054), except for TG (P = 0.362). There was also a significant interaction effect between birth weight and APOC3 , P = 0.002). No significant  interaction effect between birth weight and PPARγ2 was found.In males, low birth weight was associated with lower levels of HDLc and apoA, and higher TC/HDLc and apoB/apoA ratios Figure . Birth wAll the genotype frequencies of the analyzed gene polymorphism were in agreement with the Hardy-Weinberg equilibrium. The distribution of the APOE genotypes did not differ between males and females (P = 0.961) Table . Males wThe distribution of the APOC3 genotypes did not differ between males and females P = 0.489) . No differences were found between the Pro12/Pro and Pro12/Ala12 genotypes in any of the lipid parameters studied, nor in BMI (data not shown).There were no differences between all the genotypes in birth weight in males and females. The outcome did not change when all the analyses were adjusted for age and/or pubertal development.Low birth weight was associated with higher levels of TC, LDLc, apoB, Lp(a), TC-HDLc, TC/HDLc and apoB/apoA in males with the APOE ε3ε4 genotype, after controlling for age, pubertal status, socioeconomic status, gestational age, BMI and cardiorespiratory fitness Table . Low birLow birth weight was associated with lower levels of apoA in males with the APOC3 S1/S1 genotype, and higher levels of Lp(a) in those with both the APOC3 S1/S1 and the S1/S2 genotype. In females, low birth weight was not significantly associated with blood lipids in any of the APOC3 genotypes Table . There wThe results of the present study show that low birth weight seems to have a deleterious influence on blood lipid profile, particularly in those adolescents with the APOE ε3/ε4 genotype. The results also suggest that low birth weight is associated with lower levels of HDLc and apoA, and higher TC/HDLc and apoB/apoA ratios in male adolescents. Yet, low birth weight is not associated with blood lipid levels in female adolescents. Out of the three genes analyzed, the APOE genotypes seem to be a major determinant factor of lipid and lipoprotein profile in both sexes. Taken together, these results support the notion that intrauterine environment interact with genes in determining blood lipid profile in later life. One possible explanation to these findings is that a reduced foetal growth may have negative consequences on the liver growth. A poor liver growth may cause a down-regulation of the hepatic receptors, as well as disturbances in the synthesis of cholesterol. Consequently, the influence of the APOE genotypes on the lipid and lipoprotein metabolism would be intensified in those individuals with a low birth weight.The potential underlying mechanism linking low birth weight with and adverse lipid profile in later life remains unclear. Barker et al.  suggesteInterpretation of the birth weight effect on blood lipids after adjustment for current body size is controversial, and available results indicate that postnatal changes in size, rather than fetal growth, is important . Our finVariation in blood lipid levels seems to be partially determined by the APOE genotypes , which aFemale adolescents with the APOC3 S1/S2 genotype had elevated TC and non-HDLc levels, which is in agreement with another study . The levThe PPARγ2 gene plays an important role in the regulation of glucose, lipid and energy metabolism , therefoOther factors such as current dietary intake may have affected the lipid levels, but such data were not available in the AVENA Study. There is controversy about the influence of age at the onset of menses on lipid and lipoprotein levels. Yet, no effect of age of menarche on lipid levels have been observed in the AVENA Study population . The incThe results of this study suggest that intrauterine environment interact with the genetic background affecting the lipid profile in later life. The findings indicate that low birth weight has a deleterious influence on lipid profile particularly in adolescents with the APOE ε3/ε4 genotype.The authors declare that they have no competing interests. No benefits in any form have been received or will be received from a commercial party related directly or indirectly to the subject of this article.JRR conceived the hypothesis, conducted the statistical analyses, and drafted the manuscript. IL, FOP and LAM critically revised the drafted manuscript. DGL, AM, MGF, JAM performed the genetic analysis. All authors contributed to the interpretation and discussion of the results, and critically revised the drafted manuscript.*AVENA Study GroupCoordinator: A Marcos, Madrid.Local coordinators:MJ Castillo, Granada.A Marcos, Madrid.S Zamora, Murcia.M García Fuentes, Santander.M Bueno, Zaragoza, Spain.Granada: MJ Castillo, MD Cano, R Sola (Biochemistry); A Gutiérrez, JL Mesa, JR Ruiz ; M Delgado, P Tercedor, P Chillón , M Martín, F Carreño, FB Ortega, GV Rodríguez, R Castillo, F Arellano (Collaborators). Universidad de Granada. E-18071 Granada.Madrid: A Marcos, M González-Gross, J Wärnberg, S Medina, F Sánchez Muniz, E Nova, A Montero, B de la Rosa, S Gómez, S Samartín, J Romeo, R Álvarez,  A Álvarez  L Barrios  A Leyva, B Payá . L Martínez, E Ramos, R Ortiz, A Urzanqui. (Collaborators). Instituto de Nutrición y Bromatología. Consejo Superior de Investigaciones Científicas (CSIC). E-28040 Madrid.Murcia: S Zamora, M Garaulet, F Pérez-Llamas, JC Baraza, JF Marín, F Pérez de Heredia, MA Fernández, C González, R García, C Torralba, E Donat, E Morales, MD García, JA Martínez, JJ Hernández, A Asensio, FJ Plaza, MJ López . Dpto. Fisiología. Universidad de Murcia. E-30100 Murcia.Santander: M García Fuentes, D González-Lamuño, P de Rufino, R Pérez-Prieto, D Fernández, T Amigo (Genetic study). Dpto. Pediatría. Universidad de Cantabria. E – 19003 Santander.Zaragoza: M Bueno, LA Moreno, A Sarriá, J Fleta, G Rodríguez, CM Gil, MI Mesana, JA Casajús, Vicente Blay, María Guadalupe Blay. (Anthropometric assessment). Escuela Universitaria de Ciencias de la Salud. Universidad de Zaragoza. E-50009 Zaragoza.The pre-publication history for this paper can be accessed here:


The WHO 2006 Child Growth Standard is based on data from international optimally nourished breastfed infants from birth to age 5 years.To assess the potential effect of its use on weight and growth monitoring of UK children.Full-term members of two population-based UK birth cohorts: the Children in Focus sub-cohort of the Avon Longitudinal Study of Parents and Children (ALSPAC) (n = 1335) and the Gateshead Millennium Baby Study .Growth data from birth to 5 years were converted into z-scores relative to the WHO 2006 standard.Compared with the WHO standard, both UK cohorts had higher birth weights  and ALSPAC had higher birth lengths. After birth, length showed a good fit at all ages. By 2–4 months, both cohorts were similar in weight to the WHO median , but thereafter the UK cohorts were heavier . At age 12 months, the risk of being classified as underweight (weight <2nd centile) was considerably lower according to the WHO standard than by the UK 1990 Growth Reference , and the risk of being classified as obese at 4–5 years (body mass index >98th centile) was slightly increased .Adoption of the WHO 2006 Growth Charts would set a markedly lower standard of weight gain beyond the age of 4 months for UK infants and could support efforts to avoid future childhood obesity. However, the WHO standard is not representative of size at birth in the UK. The World Health Organisation (WHO) Child Growth Standard for infants and children up to the age of 5 years was published in April 2006. It is based on the growth of healthy breastfed children in optimal conditions between 1997 and 2003 from six different countries: Brazil, Ghana, India, Norway, Oman and USA.There is an understandable enthusiasm for the idea of adopting these charts in the UK, but before doing so it is important to assess how well UK children match, or diverge from, the new charts, in order to understand the implications for growth monitoring and clinical care. We have explored this question using data from two representative UK birth cohorts.The two datasets used were from the Gateshead Millennium Baby Study (GMS) and the Children in Focus sub-sample of the Avon Longitudinal Study of Parents and Children (ALSPAC), which between them provide detailed growth data spanning the entire period of the new charts. GMS is a prospective population-based cohort study of feeding and growth in infancy comprising 1029 babies born between June 1999 and May 2000 in Gateshead, an urban borough in the North of England. For this analysis, data from 923 full-term infants were used.The ALSPAC Children in Focus sub-cohort includes 1335 full-term infants born in Avon, south-west England, between June and December 1992. Weight and length/height measurements were collected at research clinics at birth, 4 months, 8 months, 12 months, 18 months, 24 months and 5 years.For each child, age- and sex-adjusted z-scores for weight, length (height at >2 years old) and body mass index (BMI) were calculated using exact ages at measurement by comparison with both the WHO 2006 and the UK 1990 growth data using software provided respectively by the WHO and the Child Growth Foundation . Conditional weight gain was calculated to account for regression to the mean.Both studies received appropriate ethics committee approval and obtained informed written consent from each participant.Both cohorts showed a reasonably good fit with the UK 1990 reference during the first year of life, as indicated by mean weight and length z-scores close to zero . The onlUK children had relatively high mean z-scores for birth weight and birth length compared with the WHO 2006 standard . After bBetween 4 months and 1 year, compared with the WHO standard, both cohorts showed a rapid rise in mean weight z-scores. After 1 year, the mean z-scores as assessed by the different growth reference data started to converge , fig 1.By the WHO 2006 standard, infants were considerably less likely to be classified as underweight  or having poor weight gain (downward-crossing through weight centiles) over the first year, compared with the UK 1990 Reference . ConversIn summary, adoption of the new WHO growth charts for UK children up to age 5 years would have a significant impact on the interpretation of their weight gain and growth. However, the effects are complex and appear to differ at various ages. The marked reduction in numbers of infants who would be classified as underweight or growth faltering beyond age 4 months is an expected consequence of the WHO’s decision to have the breastfed child as the normative model. However, UK infants would also be classified as being larger at birth, but not at 2–4 months, and would result in a complex pattern of weight centile changes over the first year for the average UK child .This analysis is based on data from two large representative UK birth cohorts, which between them allow comparison with the WHO charts at a wide range of ages. GMS provides detailed weight data early in infancy, and the ALSPAC provides both weight and height/length from infancy through to the pre-school years. At times of overlap, the two cohorts showed very close similarity in weights and heights, and, at least in infancy, they are also broadly similar to the UK 1990 reference. The gradual increase in weight z-scores by 4–5 years of age compared with the UK 1990 has been previously reported in ALSPAC and probably reflects the secular changes in UK children.The WHO 2006 Child Growth Standard embodies a number of novel and admirable principles, with the aim of promoting optimal infant and childhood growth. Firstly, the international MGRS source data indicated for the first time that population differences in growth are avoidable, given optimum nutrition and living conditions.The WHO published new growth charts in April 2006 based on infants of non-smoking, breastfeeding mothers living in optimal conditions in six countries.The WHO proposed that these set a standard for normal growth in infancy applicable throughout the world.At birth, UK children are longer and heavier than the WHO standard,After birth, the length of UK children matches the WHO standard closely.Use of the WHO standard would lead to far fewer UK children being classified as underweight or weight faltering in the first year, but more would be classified as overweight in the pre-school years.The WHO 2006 Growth Charts would set a lower standard of weight gain for UK infants.However, our findings, particularly during the first 2 months, suggest that these standards may not be simply transferable to the UK. On the WHO chart, UK infants would appear larger than average at birth and then cross approximately half a centile space downwards in the first few weeks of life. The explanation for this may be that, although postnatal nutrition in the WHO MGRS cohort was optimal, intrauterine growth appeared to have been constrained, as size at birth was generally smaller than in the UK. In the MGRS constituent datasets, whereas mean birth weights in Norway and USA (3.5–3.6 kg) were similar to that in the UK, the populations from several other countries showed markedly lower mean birth weights , and this appears to correlate with differences in maternal size.7The UK 1990 and other existing national growth charts do not allow for the rapid weight loss and recovery that normally occurs in the first 2 weeks of life.Beyond the first 2–4 months, use of the WHO standard would make it much less likely for UK children to be classified as underweight or growth faltering. Recent work has revealed that mild degrees of weight faltering are unlikely to be associated with major social or medical disorders,In contrast with underweight, adoption of the WHO growth chart would make UK infants and toddlers more likely to be classified as overweight or obese. There is a growing body of evidence that a higher plane of growth during infancy is associated with increased risk of obesity in children and adults.The birth weight section of the WHO chart presents other difficulties, as there is no preterm element, which is a well-used feature of UK charts. These two issues taken together suggest that it may not be desirable for the UK to adopt the birth weight section of the WHO chart, beginning its use instead after the first 2 weeks.In conclusion, the WHO 2006 Growth Standard places the breastfed child as the norm for growth. Its use would greatly reduce the numbers of UK infants classified as underweight and support efforts to avoid excess weight gain in infancy. However, the WHO 2006 Growth Standard is not representative of size at birth in the UK. In view of the resulting complex weight centile changes in the first few weeks of life, the potential confusion about feeding that this might raise with mothers, and also the absence of a preterm element to the WHO charts, the Department of Health Scientific Advisory Committee on Nutrition and the Royal College of Paediatrics and Child Health have recently jointly recommended that the WHO 2006 Growth Standard is appropriate for use in the UK children, but only from age 2 weeks.