Child undernutrition is a major public health problem in low income countries. Prospective studies of predictors of infant growth in rural low-income country settings are relatively scarce but vital to guide intervention efforts.A population-based sample of 1065 women in the third trimester of pregnancy was recruited from the demographic surveillance site (DSS) in Butajira, south-central Ethiopia, and followed up until the infants were one year of age. After standardising infant weight and length using the 2006 WHO child growth standard, a cut-off of two standard deviations below the mean defined the prevalence of stunting (length-for-age <-2), underweight (weight-for-age <-2) and wasting (weight-for-length <-2).The prevalence of infant undernutrition was high at 6 months  and at 12 months of age . Significant and consistent predictors of infant undernutrition in both logistic and linear multiple regression models were male gender, low birth weight, poor maternal nutritional status, poor household sanitary facilities and living in a rural residence. Compared to girls, boys had twice the odds of being underweight  at 6 months, and being stunted at 6 months  and at 12 months of age . Infant undernutrition at 6 and 12 months of age was not associated with infant feeding practices in the first two months of life.There was a high prevalence of undernutrition in the first year of infancy in this rural Ethiopia population, with significant gender imbalance. Our prospective study highlighted the importance of prenatal maternal nutritional status and household sanitary facilities as potential targets for intervention. There is an ongoing worldwide effort focused on the complete eradication of extreme poverty and hunger . HoweverDespite an encouraging global downward trend in the prevalence of stunting, the progress is not uniform across countries ,11 AccorEpidemiological studies conducted in developing countries have identified several factors associated with undernutrition, including low parental education, poverty, low maternal intelligence, food insecurity, maternal depression, rural residential area and sub-optimal infant feeding practices ,13,16-20In this context, the objective of our study was to describe the prevalence of undernutrition of infants aged 6 and 12 months and associated risk factors, in a population based cohort: the Perinatal Maternal Mental Disorder in Ethiopia (P-MaMiE) study .Catha edulis Forsk) and chilli-peppers are the main cash crops, while maize and "false banana" or Ensete (Ensete ventricosun) are the main staples. Due to high population density and intermittent drought, especially in the lowland area, a proportion of the population has been affected by food insecurity, leading to reliance on food aid provided through the World Food Programme.This study was conducted in the Butajira demographic surveillance site (DSS) which is located in the Meskan, Mareko and Silti Districts, Gurage Zone, in the Southern Nations, Nationalities and Peoples Regional state (SNNPR) of Ethiopia. The district capital, Butajira, is located 135 km south of the capital city, Addis Ababa, and 50 km to the west of Zway town in the Rift valley. In 2007 Butajira town had 33,393 inhabitants with a 1:1.03 male to female ratio . Currenthttp://www.butajira.org. Each DSS sub-district has one or two full-time employed enumerators. These are residents of their respective sub-districts who visit every household once every three months to collect data on vital events, allowing calculation of the exact population of the DSS.The DSS in Butajira was established in 1986 with the main objective of generating continuous and valid data on vital statistics to be used for research and interventions . The DSSHealth posts and elementary schools are located within each sub-district and are reasonably accessible . During our data collection the majority of these health posts were staffed with one male community health agent (CHA) with three to six months training in primary health care delivery.Prior to the first interview women were informed about the objective of the study and provided informed consent. Arrangements were made for the study project to pay all health related expenses of the mothers and children participating in the study. The study was granted ethical approval from the Ethics Committee of the Institute of Psychiatry, King's College London, and National Ethics Review Committee based in the Ethiopian Science and Technology Agency.This paper is part of the P-MaMiE cohort study designed to assess the predictors of short and intermediate term infant health and nutritional outcomes . The samIn six of the nine rural sub-districts, CHAs were trained to measure birth weight using SECA 725 scales measuring to an accuracy of 10 g. The CHAs live and work in the sub-district and are well-known to the women. After giving birth, participating women were requested to inform the CHA of the birth so as to enable the neonate to be weighed within 24 to 48 hours of birth. Birth weights were not measured in the remaining four sub-districts as no suitable health worker was available.A separate questionnaire was administered to women at the time of measuring birth weight or in the first few days after birth in order to obtain information on stillbirths, prolonged labour, pre-lacteal feeding, withholding of colostrum, and the timing of initiation of breast-feeding. Timing of initiation of breastfeeding was recorded using three categories  and for this analysis we combined the last two categories. In the sub-districts where birth weight was not measured, the questionnaire was administered by the project data collectors, having been informed by the traditional birth attendant or DSS enumerators that the woman had given birth. As women in this setting were unable to report their gestation at delivery with any accuracy we were unable to assess preterm delivery. It was not feasible to use last menstrual period (LMP) to estimate gestational age in this cohort because the mothers could not reliably date their LMP.Including numerous individual items in a model enables measurement of the effect of the specific detailed item on the outcome which might be helpful in designing interventions. On the negative side, there is a higher likelihood of getting false positive associations as the result of multiple tests. To overcome the latter problem without significant compromise of the first issue, we created the following four composite scores from items defining the same concept by adding responses of identified variables giving equal weights to each item. Each composite variable was used as a continuous variable:(1) Poor sanitary conditions: not having a toilet facility, not having safe water, disposing of rubbish on the field). We aggregated these three variables since all of them are known risk factors of undernutrition in Ethiopia although their internal consistency is relatively low .(2) Maternal autonomy: decision-making power reflecting whether the woman is allowed to act without asking permission from her husband . The resulting scale had a Cronbach alpha value of 0.93.(3) Availability of support to the mother: able to visit friends, enough help at home, enough help in looking after children other than the index child, enough help from husband and no violence towards her. This scale had a Cronbach alpha value of 0.47 which is relatively low. However, these items measure quite different sources of support and we would not expect them to correlate highly.(4) Poverty index: non-literate wife, non-literate husband, do not own radio, do not own bed, do not possess valuable goods like gold and jewellery, own their home, own large animals, own small animals, animals spend the night within the home, home has a window. We formed this scale by exploratory factor analysis, starting with more than 30 characteristic variables, and tested the resulting three factors with confirmatory factor analysis. The factors and their indicators were then modified to attain convergence and also meaningful factor loadings. We then defined the index as the sum of individual items with equal weight after obtaining a one factor model with meaningful factor loadings in terms of expected sign, statistical significance and width of confidence intervals. At this stage we preferred to give equal weights for the identified variables to create a composite score since we could not find convincing evidence to prioritise one over the other in terms of its potential effect on infant nutritional status in this setting. The scale had a Cronbach alpha value of 0.73 which shows an acceptable level of internal consistency.catha edulis Forsk) at least weekly (it is a widely used stimulant in the Horn of Africa and Arabian Peninsula) or drinking alcohol at least weekly), at least one obstetric complication (prolonged labour (>24 hours), assisted delivery , post-partum haemorrhage, post-partum fever), infant gender, immunisation status of infant at two months (yes/no) as an indicator of maternal health-seeking behaviour, history of infant illness in the first two months of life to the extent that the mother thought the baby was going to die (yes/no), birth weight (low (birth weight (BW) < 2500 g), normal (BW ≥ 2500 g), not measured) and early infant feeding practices. Particular focus was given to characteristics that described how the child was fed in the first two months of life: whether the infant was exclusively breast fed in the first two months as reported by the mother at the two month follow-up time-point (yes/no), if the infant was given pre-lacteal food (yes/no), if the infant was given colostrum (yes/no) and the timing of initiation of breast feeding (≤ one hour versus > one hour). We aimed to estimate the crude and adjusted effect of each of these variables upon nutritional status of infants at the ages of six and twelve months.The following characteristics were treated individually within logistic and linear regression analysis while modelling nutritional status of the infants: residence , number of children under the age of 5 years , age of father, age of mother, height of mother, mid-upper arm circumference of mother (MUAC) at recruitment, marital status (polygamous versus non-polygamous), substance use as a binary variable  estimate the fully adjusted model for all variables presented in respective tables except for the four variables listed under "feeding practices", (2) to obtain the independent effect of each feeding practice, we re-estimated the model in (1) including also the feeding practice variable of interest. This generated (a) the effect of each variable adjusted for all others except feeding practice variables, and (b) the effect of each feeding practice variable adjusted for all other variables but not for the remaining feeding practice variables. We did not adjust each infant feeding practice for the others because of collinearity and to enable us to obtain the independent effect of each feeding practice upon undernutrition. To maximise the use of available data, birth weight was included in all models as a three category variable , low birth weight (BW <2500 gm) and birth weight not recorded). All data analysis was carried out using STATA version 10 .Initial recruitment and detailed description of follow-up is presented in figure Ninety nine percent of the women were married and all younger  than their husbands  Table . The majThe prevalence of regular khat chewing during pregnancy (28%) was significantly higher than during the postnatal period (21%) . The prevalence of regular alcohol consumption was generally low among the study participants and reduced significantly during the postnatal period compared to pregnancy . In this cohort, harmful traditional practices were relatively rare with a low prevalence of performing uvulectomy (1.8% before two months of age and 4.8% before one year of age), female circumcision (1.5%) before two months of age and milk teeth extraction (3.9%) in the first year of infancy.Fifty four percent of women attended antenatal care, 90% delivered at home and only 24% of deliveries were attended by trained personnel at any level. Despite this, a high proportion of mothers reported giving colostrum to their newborn (82%), 32% initiated breastfeeding within the first hour and 16% reported giving pre-lacteal food to their new born. At the age of two months 59% of the infants had already received at least one type of vaccination: 53% Polio; 29% DPT, and 35% BCG.At one year of age 99.6% of the infants were breastfeeding. At the six month anthropometric assessment 33.4% of the mothers reported that their infants were currently ill; this prevalence was reduced to 30.6% during twelve month growth assessment. During the same follow-up 95.2% of the mothers reported at least one infant illness episode since birth, 60.0% of whom had thought that their baby was going to die because of the severity of the episode. Introduction of supplementary feeding occurred for 33.8 of infants by six months, 36.8% during 7-9 months and 28.3% after 9 months.The prevalence of infant undernutrition increased throughout the first year of life Figure . The oveCompared to the 2006 WHO child growth standards, which have a mean of zero and standard deviation of one, infants in this cohort in the first year of life were significantly lighter as reflected in their mean (95% CI) weight-for-age z score at two months , six months , twelve months  and shorter as reflected in their mean (95% CI) height-for-age z score at two months , six months  and twelve months of age .Results from bivariate analysis of each covariate and the growth outcomes are presented in Tables The adjusted effects of covariates on the infant growth outcomes are presented in Tables Linear regression using weight-for-age z score and height-for-age z score as continuous indices at both the six month and twelve month time-points did not yield any major differences from use of the dichotomised anthropometric indices (results available from the first author on request).In this study we have reported on the prevalence and predictors of infant undernutrition in a predominantly rural, population-based prospective birth cohort. Such studies are rare in resource poor countries given the lack of vital registration. Sub-optimal early feeding practices were not associated with nutritional status of infants except in unadjusted analysis where witholding colostrum and delaying initiation of breastfeeding for more than an hour were risk factors for underweight at six months of age. There was a clear and striking picture of infant undernutrition from as early as two months of age that worsened continuously to the age of one year. Girls fared better than boys implying vulnerability of the latter to the existing harsh environment. We were also able to replicate findings of important risk factors from previous studies on infant undernutrition. We found rural residence, low birth weight, poor sanitary conditions, maternal undernutrition, male gender, increased parental age and having two or more children other than index child aged under five to be the main risk factors for infant undernutrition in this study area. Scales of maternal autonomy, availability of support the mother and poverty do not seem to have a clear association with infant undernutrition in this setting, probably due to lack of variability in the study area with respect to these factors. Having older parents seems to be a risk factor for undernutrition which could possibly be explained by large family size with limited resources to be distributed among the large number of siblings.The strength of this study comes from its design and use of both binary and continuous variables to model infant nutritional status. Unlike most previous studies which were cross-sectional, we were able to measure exposure variables prospectively, free of information bias, and to assess the stability of the association across the neonatal and infant periods of development. We were also able to compare the patterns of association with the outcome of interest when analysed as binary variable (i.e. underweight and stunting) for easy practical interpretation and as a continuous variable (i.e. weight-for-age z score and height-for-age z score) to maximize statistical power. Unlike cross-sectional studies that rely on the recall of parents to determine infant age, which would be difficult in our study setting, we have calculated age based on the contemporaneously recorded date of birth. This information is not routinely available given the low percentage of women in Ethiopia who give birth in clinics or hospitals.A number of limitations can be identified that might have affected our findings. Most of the risk factors were obtained from maternal report as there was no other means of obtaining that information, and as such could have been subject to recall bias. However, the outcome was measured prospectively and so maternal recall bias is less likely to have affected the observed associations. Several research workers were involved in measuring height and weight outcomes, which might have introduced measurement errors. Inter-rater reliability was not assessed formally. The training, close follow-up by first authors (GM and CH), periodic quality-control checks and re-measuring of any suspected error cases within one week, should have minimised measurement errors. Any resulting error is likely to be random, and would have had the effect of reducing the size of any genuine effect towards the null. Only measuring birth weight in selected sub-districts might have influenced some of the associations, although non-measured birth weights were missing at random . To accoThe prevalence of undernutrition increases progressively in Ethiopia from early infancy, peaking in the late months of the second year of age and thereafter starting to stabilise. The prevalence of stunting starts to exceed that of underweight when undernutrition starts to stabilise . The curLike other African countries  breastfeMaternal MUAC as a measure of nutritional status in pregnancy had a significant protective effect upon an infant being underweight at both six and twelve month assessments, even after adjusting for possible risk factors, but was not associated with stunting at either time-point. In a cross-sectional sample of infants aged 5-11 months in two rural villages of Ethiopia, maternal height, triceps skin fold thickness and zinc concentration in beast milk were associated with stunting but current weight and MUAC were not associated with stunting . MaternaIn this study, the mean poverty index score was significantly higher for mothers whose infants were undernourished compared to mothers whose infants were not. However, the mean maternal autonomy score and mean score of availability of support to the mother did not differ between the two groups of women. In the adjusted model none of the three composite scores was significantly related to infant undernutrition. As the primary caregiver, it is probable that maternal autonomy is critical for the overall wellbeing of infants  althoughLow birth weight was a significant predictor of stunting and underweight at twelve months and underweight at six months of age. This is in line with a previous study from Pakistan reporting a higher risk of stunting at one year of age for low birth weight infants . In the Living in poor sanitary conditions significantly increased the risk of an infant being underweight and stunted at 12 months of age and significantly predicted underweight at six months of age. Although these findings might be expected, improving sanitation is an important area for increased intervention to reduce undernutrition . Other sIn the present study boys were more likely to be underweight at six months and more likely to be stunted both at six and at twelve months. A significantly higher prevalence of undernutrition amongst boys as compared to girls is consistent with other Ethiopian studies ,15,58 anRural residence significantly predicted underweight and stunting at twelve months of age in line with other African countries . This waHaving one previous child aged less than 5 years was found to be protective of being underweight at six months of age which might be explained by the child caring experience of the mother with her first child. There was a statistically non-significant but consistent association with underweight and stunting at six and twelve months of age showing an increased risk of undernutrition with an increased number of under five children prior to index child. A regional survey within Ethiopia  and a stIn our study infant undernutrition was common and progressively worsened throughout the first year of life. After adjusting for other risk factors, none of the measured early infant feeding practices were associated with undernutrition, suggesting that interventions directed only at these maternal behaviours may have a limited impact in this setting. However, interventions targeted at improving maternal nutritional status during pregnancy might reduce infant undernutrition by increasing birth weight and improving the quantity and quality of breast milk. It might be possible to minimise the rural-urban difference in infant undernutrition by empowering mothers living in rural areas with skills of optimal parenting practices and increasing accessibility of health services. Investment in improving basic sanitary conditions and devising mechanism to improve maternal under-nutrition should be the focus for short-term strategies to reduce infant undernutrition in this setting.The authors declare that they have no competing interests.CH, MP, AA conceived the idea; CH designed the study; GM and CH coordinated data collection and data entry; GM analysed the data and drafted the manuscript, CH, MD, FT, BW, MP, MT, critically commented on the draft manuscript, all authors contributed to the interpretation of results and approved the final manuscriptThe pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2458/10/27/prepub


The patient usually presents with a slowly enlarging asymptomaticswelling. They are usually diagnosed in early stages because of cosmetic problems. In ourpaper we report a  Nasolabial cysts are very rare nonodontogenic soft-tissue lesions of nasal vestibule, fossa canina, and sublabial region. The lesions cause painless swelling in sublabial fold, lips, face and cause nose obstruction. Pain can occur if the cyst becomes infected. The incidence of the cyst is 0.7% in overall chin cysts . The ini Herein we present diagnosis, differential diagnosis and treatment of a nasolabial cyst in the lights of the literature.A 53-year-old man was admitted to the ENT Department. He complained of a firm mass in the right nasolabial area that expands lips outwards. He had a history of having tooth extraction a month ago. The past medical history was unremarkable.  On otolaryngologic examination, palpation revealed a swollen area corresponding to the anatomic location of the nasolabial fold. The swollen area was palpated intraorally, it was firm and nontender. The floor of nose was swollen and narrowed by a mass from the inferior side. The lesion was 2 × 3 cm in dimensions . The CT The nasolabial cyst was excised via a sublabial intraoral incision under general anesthesia. The cyst was dissected free, the wound was closed in layers. There were no postoperative complications. The histopathological examination revealed nasolabial cyst. The symptoms resolved after the operation.Nasolabial cysts were first described by Zuchercandl in 1892 .There are three theories for the formation of the cyst.The cyst is formed embryologically by detention cells in the maxilla, medial, and lateral nasal wall.The cyst is formed embryologically by detention cells from the inferior nasolacrimal channel redundant cells.The cyst is formed embryologically by detention cells from the inferior nasolacrimal channel endodermal cells . Exposurbase of the nose, and neoplasms of base of the nose [examination and histopathological examination. The differential diagnosis of the cyst must be made with central line cysts, cyst of maxilla, odontogenic cysts, periapical cysts, periapical abscess, periapical granulomas, epidermal inclusion cyst, frunculosis of the nose . The safthe nose . The dia The treatment can be made by surgical exicion, injection of sclerozing materials in the cyst, and endoscopic marsupialization methods . Exicion Nasolabial cyst must be kept in mind in differential diagnosis of nasal vestibule, nasal base, and sublabial area . The dia


The control group without malignant disease was established by random sampling from the same source population as the cases (n=108 209). Data on birth weight, gender, mother's residence and parental age at the time of birth were collected from the Medical Birth Registry of Norway and data on cancer from the Cancer Registry of Norway. The Mantel–Haenszel test of linear trend showed no trend in risk across the birth weight categories: individuals in the highest quartile of birth weight (⩾3860 g) had an odds ratio (OR) of 1.19  compared to individuals with birth weight <2500 g. The adjusted OR was 0.81 (95% CI: 0.52–1.26) for birth weight below 2500 g (exposed). Though not statistically significant, the results suggest that low birth weight might influence the risk of melanoma later in life.We investigated whether lower birth weight was associated with lower risk of melanoma later in life. This population-based case–control study included all incident cases of histologically verified invasive melanoma diagnosed until 31 December 2003 in the Norwegian population born between 1967 and 1986 ( Exposure to intermittent high-dose ultraviolet (UV) radiation is widely recognised as the major cause of cutaneous malignant melanoma (hence referred to as melanoma). In two previous studies, we observed a lower incidence of melanoma in twins as compared to singletons  for singletons and 2367 g for twins (It has been hypothesised that birth weight can influence the cancer risk later in life. This possible association has been studied for melanoma (We investigated whether lower birth weight is related to lower risk of melanoma later in life by estimating the association between birth weight and the incidence of melanoma in the Norwegian population.n=1 142 635) in Norway between 1 January 1967 and 31 December 1986. From this 20-year cohort, a case–control study was carried out.This population-based study includes all live-born men and women (n=859). Melanoma was defined according to the international classification of diseases, and only cases with histologically verified invasive melanoma of primary tumour were included in the study (n=711). The control group is a random sample of the population, without a malignant disease, born in the same period of time as the cases and in which the individuals were all alive at the closing date 31 December 2003 (n=109 727).All incident cases of melanoma diagnosed in that population before 31 December 2003 were identified . This is a population-based registry based on compulsory notification of every birth or late abortion with a gestational age of 16 weeks or more. Infants and parents can be identified thorough the national 11-digit personal identification number .Data on cancer were collected by linkage to the Cancer Registry of Norway, using the unique 11-digit identification number. The cancer registry is population-based .n=87) and cases of multiple cancers (n=61) were excluded from the analyses. Individuals lost to follow-up  and individuals with follow-up shorter than 1 year (due to emigration or death) were also excluded . This 1-year limit was based on the definition of infant mortality rate  were also excluded. The final study population includes 709 cases of melanoma and 108 209 controls without a malignant disease.Cases without histological verification of the primary tumour (ity rate . FurtherThe birth weight (g) was analysed both as a continuous variable and as a categorical variable: <2500 and ⩾2500 g. The two categories were defined by the low birth weight (LBW) limit of 2500 g . Birth wNorway is geographically spread from 58 to 72° latitude north and the variation in UV radiation is large. The variable ‘place of birth’, with the categories ‘North Norway’ and ‘South Norway’, was created using the separation line located approximately at 62° latitude north.n=32 and controls, n=8305).The variables ‘mother's age’ and ‘father's age’ were polytomised into quartiles, and the 75% quartile limits were used to create two levels in each variable: younger and older parents. We have missing data on father's age  was used to pinpoint effect modification via the Breslow–Day test of heterogeneity. Quantification of confounding was done by comparing the crude estimate of OR and the adjusted Mantel–Haenszel OR. A logistic multivariable model was used to control for multiconfounding  and the disease (melanoma) was the major focus . The crut-test showed no statistically significant difference between the case and the control groups. Similarly, no difference was found when parental age was assessed as a categorical variable  in relation to men. The risk of melanoma was 30% higher  for individuals born in the south, as compared to individuals born in the north . Evaluatvariable . MaternaThe controls had a lower mean birth weight, regardless of whether the genders were combined or separated for analysis . HoweverIn the Mantel–Haenszel test of linear trend, the individuals in the highest quartile of birth weight (⩾3860 g) had an OR of 1.19 (95% CI: 0.77–1.84) compared to individuals in the lowest category (<2500 g) . There wResults from the stratified analysis, estimating the risk of melanoma associated with birth weight, independent of gender, showed the adjusted Mantel–Haenszel OR of 0.84 (95% CI: 0.55–1.27), which corresponds to a confounding effect of –2.3%. The confounding effect for each of the other variables in the analysis  was below 5%. No effect modification was observed (data not shown).A power calculation was performed for the effect of OR=0.81, considering the frequency of exposition to birth weight <2500 g in controls to be 3.7% , with a We questioned whether individuals with a LBW were less likely to develop cutaneous malignant melanoma compared to individuals with higher birth weight.Results from the univariable and the multivariable analyses indicate that birth weight might be associated with occurrence of melanoma, also when other possible confounders, such as gender, place of birth and parental age were included in the model.In 2001, The observed tendency to reduced melanoma risk among twins agrees with the tendency to lower risk among those with birth weight below 2500 g, as the mean birth weight for twins was 2640 g. Maternal preeclampsia/eclampsia, hypertension during pregnancy and chronic kidney disease may influence the nutrition conditions in the womb and often lead to undernutrition of the foetus. These conditions were more frequent among controls than in the case group. These observations might support our hypothesis that LBW can reduce the risk of melanoma later in life.In the present study, women have close to 2.5-fold risk of developing melanoma compared to men. This is possibly due to the different sun exposure habits between genders, but other factors might also be involved. Individuals born in the south of Norway had a 30% higher risk of melanoma compared to individuals born in the north. This result is due to the latitude gradients in UV exposure in Norway, which previously has been shown to be reflected in the melanoma risk . No effeResults from the present study indicate that LBW (<2500 g) might be associated with lower risk of melanoma later in life. Birth weight is probably not a risk factor of melanoma, but it might be a marker of other relevant factors. Investigation of other exposures correlated with birth weight should therefore be investigated in the future.


P-value for linear trend=0.03). This corresponded to an RR of 2.31  per 1 kg increase in birth weight. Birth weight was also a predictor of postnatal growth, that is, women who were heavy at birth remained taller and heavier throughout their childhood and young adulthood. However, the effect of birth weight on premenopausal breast cancer risk was only reduced slightly after simultaneous adjustment for height and body mass index (BMI) at age 2 years and height and BMI velocities throughout childhood and adolescence  per 1 kg increase in birth weight). The pathways through which birth weight is associated with premenopausal breast cancer risk seem to be largely independent of those underlying the relation of postnatal growth to risk.Several studies have found positive associations between birth weight and breast cancer risk at premenopausal ages. The mechanisms underlying this association are not known, but it is possible that it may be mediated through childhood growth. We examined data from a British cohort of 2176 women born in 1946 and for whom there were prospective measurements of birth weight and of body size throughout life. In all, 59 breast cancer cases occurred during follow-up, 21 of whom were known to be premenopausal. Women who weighed at least 4 kg at birth were five times (relative risk (RR)=5.03; 95% confidence interval=1.13, 22.5) more likely to develop premenopausal breast cancer than those who weighed less than 3 kg ( Positive linear relationships between birth weight and breast cancer risk, particularly at premenopausal ages have been reported . Self-reported data on breast cancer diagnosis were collected through the various follow-up contacts. In addition, in 1971, when the National Health Service Central Register (NHSCR) started to record cancers occurring in the UK population, all cohort members (including those with whom there was no longer direct contact) were ‘flagged’ at the NHSCR to provide notification of cancers, deaths and emigrations for the cohort.The National Survey of Health and Development is a socially stratified cohort of 5362 single legitimate live-births that occurred in Britain during the week 3–9 March 1946 and who have since been followed up. Most of the follow-up contacts were home interviews . These fOf the 2547 women in the birth cohort, 2176 (85%) who were known to be still alive on 1 January 1971 and for whom there was birth weight information and at least one height measurement in childhood, were included in the present analysis. Of the 371 who were excluded, 114 (31%) had died from nonbreast cancer causes and 176 (47%) had emigrated before 1971 and were no longer in contact with the study. A further 70 (19%) did not have any height measurements between ages 2 and 15 years and 11 (3%) had no birth weight data.Approval for the study was obtained from all relevant ethics committees.Breast cancer relative risks (RRs) by birth weight were estimated as rate ratios using a Cox regression model , where aBirth weight was analysed both as a continuous and as a categorical variable . To assess whether the effect of birth weight on breast cancer incidence was mediated through postnatal growth, we fitted models that included birth weight plus height and BMI throughout childhood and adolescence (at ages 2–15 years) and height at age 36 years, used as a proxy for the final height achieved at the end of the adolescent growth spurt. These anthropometric variables were specified either as age-specific attained values (reflecting cumulative growth up to a given age) or as velocities between consecutive ages . For some cohort members, childhood anthropometric data were not available at all follow-up contacts and so a multiple imputation procedure  was usedP=0.24) or occurrence of breast cancer (P=0.67).In all, 59 breast cancer cases occurred during the follow-up period. The median age at incidence was 49 years (25th and 75th percentiles: 45 and 52 years). The median age at last follow-up for the remaining 2117 women was 53 years . A total of 21 cases were known to be premenopausal (with ages of diagnosis ranging from 36 to 51 years) and nine were known to be natural postmenopausal at the time of diagnosis of breast cancer (age at diagnosis: 46–53 years). A further 12 had undergone hysterectomy before diagnosis or reported to have been postmenopausal at the time of diagnosis but did not indicate whether the menopause was natural. No information on menopausal status was available for the remaining 17 women. Anthropometric measures in childhood were available at each age between 76 and 90% and in adulthood for about 75% of all eligible women in the cohort. The probability that a measure was missing was unrelated to the woman's values at later ages or to her subsequent risk of breast cancer. For instance, having missing height data at age 7 years was not associated with adult height . Women who weighed at least 4 kg at birth were five =1.13, 22.5) times more likely to develop premenopausal breast cancer than those who weighed less than 3 kg. This corresponded to a rate ratio of 2.37  per 1 kg increase in birth weight. The effect of birth weight on all-ages breast cancer incidence was in the same direction, but was of a much smaller magnitude (P=0.02). In contrast, adult BMI (at age 36 years) was inversely related to premenopausal breast cancer rates (33% reduction in rates per every s.d. (about 4 kg m−2) increase in BMI), but the association was not statistically significant (P=0.20). Similar relationships were observed with all-ages breast cancer.Birth weight was also a predictor of height and BMI trajectories throughout childhood and young adulthood , using data derived from the multiple imputation procedure, showed that simultaneous adjustment for attained height at age 2 years and height velocities thereafter decreased the RR (as estimated by odds ratio from a logistic regression model) associated with 1 kg increase in birth weight from 2.31  to 1.87 , whereas simultaneous adjustment for attained BMI at age 2 years and subsequent BMI velocities had little impact on the magnitude of the birth weight effect ). When both height and BMI at age 2 years and their interval-specific velocities were taken into account simultaneously, the magnitude of the birth weight effect decreased only slightly from its original value of 2.31 to 1.94 . Similar results were obtained using only the subsets of women for whom the relevant measurements were available, although they were based on much smaller numbers   and 1.50 , respectively). Similar results were obtained when the analyses were restricted to the subset of women for whom measurements were available at all follow-up contacts.n=720)  and 2.20 , respectively). Similarly, the magnitude of the birth weight effect did not change with further adjustment for age at menarche in the group of women with known age at menarche (n=766)  and 2.78 , respectively). Further adjustment for maternal age, birth order, father's social class, age at first birth, parity or adult BMI provided similar results.Further adjustment for adult leg length, a better marker of growth in childhood than adult height, did not affect the magnitude of the birth weight effect on premenopausal breast cancer risk in the subset of women for whom this measure was available  age-specific measurements should have provided a more accurate measure of the underlying growth trajectory for each woman. No measurements were taken between birth and age 2 years, but further adjustment for the difference between a girl's standardised rank in height (or BMI) at age 2 years and her standardised rank in birth weight, this difference being taken as an indicator of early growth, did not alter the reported effects of birth weight on breast cancer risk. Although the available anthropometric data did not allow calculation of age at peak height velocity, the analysis took into account velocity of growth in early childhood and age at menarche, factors known to be closely related to age at peak height velocity (in utero exposure to high concentrations of these hormones may affect the mammary tissue and the risk of malignancy later in life in the offspring (It has been hypothesised that the observed associations between birth weight and breast cancer risk may reflect intrauterine exposure to oestrogens and other biological factors that may increase the number of stem cells in the breast gland and/or their proliferation , 2003. Pffspring , 2003. BIn short, the findings from this study suggest that the relationship between birth weight and premenopausal breast cancer risk is largely independent, rather than a correlate of the association of childhood growth and risk of this tumour.


Mediators of myocardial inflammation, predominantly cytokines, have for many years been implicated in the healing processes after infarction. In recent years, however, more attention has been paid to the possibility that the inflammation may result in deleterious complications for myocardial infarction. The proinflammatory cytokines may mediate myocardial dysfunction associated with myocardial infarction, severe congestive heart failure, and sepsis. A growing body of literature suggests that inflammatory mediators could play a crucial role in ischemia–reperfusion injury. Furthermore, ischemia–reperfusion not only results in the local transcriptional and translational upregulation of cytokines but also leads to tissue infiltration by inflammatory cells. These inflammatory cells are a ready source of a variety of cytokines which could be lethal for the cardiomyocytes. At the cellular level it has been shown that hypoxia causes a series of well documented changes in cardiomyocytes that includes loss of contractility, changes in lipid metabolism and subsequent irreversible cell membrane damage leading to cell death. For instance, hypoxic cardiomyocytes produce interleukin-6 (IL-6) which could contribute to the myocardial dysfunction observed in ischemia reperfusion injury. Ischemia followed by reperfusion induces a number of other multi-potent cytokines, such as IL-1, IL-8, tumor necrosis factor-α (TNF-α), transforming growth factor-β1 (TGF-β1) as well as an angiogenic cytokine/ growth factor, vascular endothelial growth factor (VEGF), in the heart. Intrestingly, these multipotent cytokines (e.g. TNF-α) may induce an adaptive cytoprotective response in the reperfused myocardium. In this review, we have included a number of cytokines that may contribute to ventricular dysfunction and/or to the cytoprotective and adaptive changes in the reperfused heart.


Perfluorooctane sulfonate (PFOS) and perfluorooctanoate (PFOA) are man-made, ubiquitous, and persistent contaminants in the environment, wildlife, and humans. Although recent studies have shown that these chemicals interfere with fetal growth in humans, the results are inconsistent.Our goal was to investigate the correlation between relatively low levels of PFOS and PFOA in maternal serum and birth weight and birth size.We conducted a hospital-based prospective cohort study between July 2002 and October 2005 in Sapporo, Japan. A total of 428 women and their infants were involved in the study. We obtained characteristics of the mothers and infants from self-administered questionnaire surveys and from medical records. We analyzed maternal serum samples for PFOS and PFOA by liquid chromatography–tandem mass spectrometry (LC/MS/MS).10 unit: β = −148.8 g; 95% confidence interval (CI), −297.0 to −0.5 g]. In addition, analyses stratified by sex revealed that PFOS levels negatively correlated with birth weight only in female infants . However, we observed no correlation between PFOA levels and birth weight.After adjusting for confounding factors, PFOS levels negatively correlated with birth weight [per login utero exposure to relatively low levels of PFOS was negatively correlated with birth weight.Our results indicate that  Perfluorinated chemicals, which have been manufactured for > 50 years, have been used in a wide range of industrial and consumer products. Perfluorooctane sulfonate (PFOS) and perfluorooctanoate (PFOA), which are representative of perfluorinated chemicals, have recently been found to be widespread contaminants in the environment, wildlife, and humans . The worPFOS and PFOA contamination in human blood has been reported in various countries . MaternaExposure of pregnant rats and mice to PFOS led to a reduction in birth weight , and a sHuman studies have shown no substantial changes in hematologic, lipid, hepatic, thyroid, or urinary characteristics in populations exposed occupationally to perfluorinated chemicals , 2003. PGiven that few reports have actually suggested a correlation between prenatal non-occupational PFOS and PFOA exposure and reduced fetal growth, further research is needed to clarify this potential relationship. The aim of the present study was to investigate the relationship between relatively low levels of PFOS and PFOA in maternal serum and birth weight and birth size, including length, chest circumference, and head circumference.We performed a prospective cohort study between July 2002 and October 2005 at the Sapporo Toho Hospital in Hokkaido, Japan . Study subjects were women who enrolled at 23–35 weeks of gestation during a routine gynecologic checkup and delivered at the Sapporo Toho Hospital. All subjects were native Japanese and residents of Sapporo or surrounding areas. Of 1,796 potentially eligible women, 514 agreed to participate (a participation rate of 29%). Among the 514 pregnant women who composed the initial cohort, 10 were excluded because of miscarriage, stillbirth, relocation, or voluntary withdrawal from the study before follow-up.The subjects completed a self-administered questionnaire survey after the second trimester of pregnancy. The questionnaire included information related to dietary habits, smoking status, alcohol intake, caffeine intake, household income, and educational level. We obtained information on dietary habits during pregnancy, such as intake of eggs, milk and its by-products, carbohydrates, fish, and meat, from responses on the food frequency questionnaire, divided into five categories: rarely/never, one to two times per month, one to two times per week, three to four times per week, or every day. In this study, we based maternal smoking status on maternal self-reporting and categorized it into two groups: nonsmokers [those who did not smoke throughout pregnancy or quit smoking at the beginning of pregnancy (first trimester)] and smokers .For estimating alcohol intake, we used the modified self-administered questionnaire described previously by For estimating caffeine intake, we used the modified self-administered questionnaire described previously by Annual household income based on maternal self-reporting was divided into five categories: < 3 million yen, 3–5 million yen, 5–7 million yen, 7–10 million yen, or ≥ 10 million yen. Maternal educational level based on maternal self-reporting was divided into four categories: 9 years, 10–12 years, 13–16 years, or ≥ 17 years. We obtained the information for exclusion criteria and potential confounders of the mothers and their infants—prepregnancy body mass index (BMI), pregnancy complications, gestational age, infant sex, parity, infant disease, birth weight, and birth size—from medical records written by gynecologists. This study was conducted with written informed consent from all subjects and was approved by the institutional ethical board for epidemiologic studies at the Hokkaido University Graduate School of Medicine.A 40-mL blood sample was taken from the maternal peripheral vein after the second trimester of pregnancy. When this was not possible because of anemia of the mother, we took a blood sample after delivery. All samples were stored at −80°C until analysis. The analytical detection method used is the variation of a method published previously , 2005b, We obtained PFOS  and PFOA  from Fluka Chemie AG . We purchased perfluoroheptanoic acid , which we used as an internal standard, from Sigma Chemical Co. . We purchased acetonitrile and methanol for high-performance liquid chromatography and analytical-grade ammonium acetate and acetic acid from Wako Pure Chemical Industries, Ltd. . The water purification system was a Milli-Q gradient A 10  equipped with an EDS polisher (activated carbon filtration) .n = 6), and the recoveries for spiked 1 and 10 ng/mL PFOA were 100.0%  and 97.3% , respectively (We performed liquid chromatography–tandem mass spectrometry (LC/MS/MS) with a Waters Quattro micro system . Separation was achieved on an Inertsil ODS-3 column  equipped with a Mightysil RP-18 GP precolumn . The column oven was maintained at 40°C. The column-switching LC/MS/MS was coupled with an online extraction system consisting of the LC/MS/MS combined with an LC pump  and a Waters Oasis HLB extraction column . After we injected a 50-μL sample with an autosampler, we loaded it onto the extraction column by flowing 50 mM acetate buffer (pH = 4.7):methanol  at a flow rate of 1.0 mL/min. The valve was switched 5 min after sample injection. We eluted the sample through a back-flashing extraction column to the analytical column and introduced it into the MS/MS apparatus. We carried out separation using a mobile phase of 1.0 mM ammonium acetate/acetonitrile (vol/vol) at a flow rate of 0.2 mL/min. The gradient profile was as follows: linear increase from 45% to 85% acetonitrile solution for 5–12 min, then hold at 85% for 3 min. The conditions for MS/MS were as follows: Desolvation and source temperatures were set at 350°C and 100°C, respectively, and the capillary was held at a potential of 600 V relative to the counterelectrode in the negative-ion mode for all compounds. Cone and desolvation gas flow rates were 50 and 350 L/hr, respectively. Cone and collision voltages were 60 and 65 V for PFOS, 30 and 18 V for PFOA, and 28 and 18 V for PFHpA, respectively. When we tested the recoveries using human plasma samples, the recoveries for spiked 5 and 50 ng/mL PFOS were 99.3%  and 97.5% , respectively  with 0.2 mL internal standard solution containing acetonitrile, centrifuged the mix at 1,450 × g for 10 min, and transferred the supernatant to a polypro-pylene tube. We subjected the aliquot of the filtered sample solution to column-switching LC/MS/MS.p-dioxins, 10 polychlorinated dibenzofurans, and 4 non-ortho-polychlorinated biphenyls (PCBs) using high-resolution gas chromatography/high resolution mass spectrometry (HRGC/HRMS) equipped with a solvent cut large-volume (SCLV) injection system  and we also measured 8 mono-ortho-PCBs and 58 non-dioxin-like PCB congeners in maternal blood (n = 370) using HRGC/HRMS at Fukuoka Institute of Health and Environmental Sciences. The analytical method is available elsewhere  in the fully adjusted model as a potential predictor of head circumference. When we examined the levels of PFOS and PFOA in samples taken at different times (during pregnancy and after delivery) by the Mann–Whitney U-test, we found significant differences in the levels of PFOS (p < 0.001) and PFOA (p < 0.001). Thus, we also adjusted the blood sampling time in the fully adjusted model. We performed all statistical analyses using SPSS for Windows, version 13.0J . Results were considered statistically significant if p < 0.05.We analyzed correlations between PFOS and PFOA concentrations and characteristics of mothers and infants by the Spearman correlation test, Mann–Whitney p < 0.05) in median PFOS concentrations by age, smoking status, blood sampling period, type of delivery, gestational age, and parity, and we observed differences in median PFOA concentrations by caffeine intake, blood sampling period, gestational age, infant sex, and parity. n = 428), and p < 0.01).t-test and Pearson’s correlation test. The mean birth weight, chest circumference, and head circumference were significantly lower in the smoker group compared with nonsmoker group (p < 0.05). Gestational age (days) had the greatest positive impact on infant birth weight and birth size . Maternal prepregnancy BMI positively correlated with birth weight (p < 0.05). Length and head circumference in male infants were significantly greater compared with female infants . Head circumference for first infants was significantly smaller than for subsequent infants (p < 0.05).10-transformed PFOS and PFOA concentrations. Multivariate models of birth weight and birth size are adjusted for gestational age only or fully adjusted; the fully adjusted model includes risk factors correlated with birth weight and birth size at p-values < 0.05, risk factors known to be correlated with birth weight and birth size from previous reports, and blood sampling period . In the crude model, we found no significant correlations between PFOS and PFOA levels and birth weight and birth size. In the model adjusted only for gestational age, a log10-unit increase in PFOS and PFOA levels correlated with a decrease in mean birth weight of 148.7 g  for PFOS and 119.1 g  for PFOA. Further adjustment for parity resulted in a loss of the significant negative correlation between PFOA levels and birth weight, so the fully adjusted model showed no significant negative correlation between PFOA levels and birth weight. In contrast, a log10-unit increase in PFOS levels correlated with a decrease in mean birth weight of 148.8 g  for PFOS in the fully adjusted model. 10-transformed maternal PFOS concentrations.10-unit increase in PFOA levels correlated with a decrease in mean birth weight of 166.4 g . Further adjustment for parity resulted in a loss of significant negative correlation between PFOA levels and birth weight in male infants, so the fully adjusted model shows no significant negative correlation between PFOA levels and birth weight in male infants. In the crude model, a log10-unit increase in PFOS levels correlated with an increase in mean length of 1.230 cm . However, after adjustment for gestational age, the significant positive correlation disappeared.10-unit increase in PFOS levels correlated with a decrease in the mean birth weight of 237.1 g . In the fully adjusted model, a log10-unit increase in PFOS levels correlated with a decrease in mean birth weight of 269.4 g . 10- transformed PFOS concentrations. We observed marginally significant negative correlations between PFOS levels and length and chest circumference in the fully adjusted model.n = 428), we measured the levels of total dioxins in maternal blood  (n = 318) and the levels of total PCBs (sum of 70 PCB congeners (58 non-dioxin-like PCB congeners and 12 dioxin-like PCBs) in maternal blood (n = 318) (data not shown). Total dioxins levels positively correlated with PFOS and PFOA levels . Furthermore, total PCB levels positively correlated with PFOA levels  but not with PFOS levels . However, the fully adjusted regression coefficients between PFOS and PFOA levels and birth weight also changed little when we adjusted for total dioxins and total PCB levels (data not shown).Among the subjects analyzed in this study . However, after adjusting for parity, blood sampling period, smoking status during pregnancy, and maternal age in a multiple regression analysis, the significant correlations between log10-transformed PFOS levels and type of delivery and gestational age disappeared . Thus, the correlations between log10-transformed PFOS levels and type of delivery and gestational age might have been confounded by parity, blood sampling period, smoking status during pregnancy, and maternal age. Further, after adjusting parity and blood sampling period in a multiple regression analysis, the significant correlations between log10-transformed PFOA levels and caffeine intake and gestational age also disappeared . Thus, the significant correlations between log10-transformed PFOA levels and caffeine intake and gestational age might have been confounded by parity and blood sampling period. On the other hand, the correlations between log10-transformed PFOA levels and sex were not significant after adjustment for parity and blood sampling period (p = 0.071). In addition, besides the covariates in our final model, we evaluated the caffeine intake as a possible confounding factor; this adjustment did not materially change the fully adjusted regression coefficients between PFOS and PFOA levels and birth weight (data not shown).Concentrations of PFOS and PFOA decreased with increasing parity, which is consistent with two previous reports . This miOur study constitutes a prospective cohort study, which minimizes recall bias. We collected prenatal information on the mother and child, such as disease history, pregnancy conditions, and birth weight and birth size, from medical records written by gynecologists, not from maternally reported information, adding data reliability. Besides the covariates in our final model, we evaluated the frequency of food intake , season of blood sampling, and season of birth as possible confounding factors. These adjustments did not materially change the fully adjusted regression coefficients between PFOS and PFOA levels and birth weight (data not shown). This study has some limitations. First, our sample size was relatively small compared with the study by This study shows a correlation between maternal PFOS levels and reduced fetal growth. PFOS cannot be degraded by the environment, and therefore we must evaluate its potential toxicity in humans, especially children. Other than birth weight, animal studies have shown effects of PFOS and PFOA on numerous organ systems, thyroid function, endocrine system, immune function, and congenital anomalies . Little p-values were incorrect in the manuscript originally published online; they have been corrected here.In the third-to-last paragraph, the Spearman rank correlation coefficients and


Malaria has a negative effect on pregnancy outcome, causing low birth weight, premature birth and stillbirths, particularly in areas with high malaria transmission. In Rwanda, malaria transmission intensity ranges from high to nil, probably associated with variable altitudes. Overall, the incidence decreased over the last six years (2002–2007). Therefore, the impact of malaria on birth outcomes is also expected to vary over time and space.Obstetric indicators (birth weight and pregnancy outcome) and malaria incidence were compared and analyzed to their association over time (2002–2007) and space. Birth data from 12,526 deliveries were collected from maternity registers of 11 different primary health centers located in different malaria endemic areas. Malaria data for the same communities were collected from the National Malaria Control Programme. Associations were sought with mixed effects models and logistic regression.In all health centres, a significant increase of birth weight over the years was observed (p < 0.001) with a significant seasonal fluctuation. Malaria incidence had no significant effect on birth weight. There was a slight but significant decreasing effect of malaria incidence on the occurrence of premature delivery  and still birth . Altitude showed a slight but significant negative correlation with birth weight. Overall, a decrease over the years of premature delivery (p = 0.010) and still birth (p = 0.036) was observed.In Rwanda, birth weight and pregnancy outcome are not directly influenced by malaria, which is in contrast to many other studied areas. Although malaria incidence overall has declined and mean birth weight increased over the studied period, no direct association was found between the two. Socio-economic factors and improved nutrition could be responsible for birth weight changes in recent years. Plasmodium falciparum malaria is an important contributor to low birth weight, maternal morbidity, preterm delivery, perinatal morbidity and mortality (still births) [P. falciparum infection during pregnancy is estimated to cause as many as 10,000 maternal deaths each year, contributing to approximately 2 – 15% of maternal anaemia, 8 – 14% of infants with LBW , and 3 – 8% of all infant deaths [Plasmodium and the consequent inflammatory response is the cause for this pathology. Primigravid and secundigravid women have the highest risk of placental parasitaemia and malaria-associated LBW [In sub-Saharan Africa, over 50 million women living in malaria endemic areas become pregnant each year  and Plas births) . In areat deaths . The invated LBW ,8-10. Itated LBW ,12.Plasmodium falciparum prevalence up to 80%; 2) 780 to 1,000 m above sea level with a P. falciparum prevalence varying between 30% to 50%; 3) the central plateau, situated at an altitude of 1,500 m to 1,800 m above sea level with a P. falciparum prevalence of 5% to 15%; 4) The Rusizi – Tanganyika plateau at an altitude of 1,800 to 3,000 m above sea level with a P. falciparum prevalence of 0 to 2%. It gives an indication of the differences between different regions but lacks details and does not take into account the variation in altitude within the transmission zones. Incidence data are widely available by the detailed reporting system of the curative health services. The PNLP and others [Most of the presented studies that associate maternal malaria and LBW use data from studies that compare individual patients in high transmission zones, where chronic parasitaemia is prevalent. Studies in low transmission areas are scarce as is information of the effect of low birth weight in Rwanda. Rwanda changed from a country with high placental malaria burden with 75% of the studied placenta's harboring parasites in 1998  to an avst to the 15th of each month from January 2002 until December 2007 in different malaria endemic regions in Rwanda. Two types of existing data sources were used: birth registries of primary health centers (PHC) and malaria data sources from the PNLP. Ethical approval for this study was sought and obtained from the Rwanda National Ethics Committee (IRB1497)This was a retrospective, observational study comparing obstetric data from all pregnancy outcomes recorded from the 1Data were collected from 11 PHCs in Rwanda that represent the four different transmission zones in the Rwanda  or as presumed malaria  and recorded as such together with the age of the patients: < 1 year, 1–4 years, 5–14 years, and 15+ years. For the present study, the aggregated monthly data, i.e. the numbers of confirmed and presumed malaria and the totals, were used.All Rwandan PHCs record all births, birth outcomes and birth weights in specific birth registers. The birth registers from the 11 study PHCs were explored as the source for birth data in the present study. The information on births and outcome recorded during the first 15 days of every month from 2002 until 2007 were copied anonymized to case report forms (CRF). Collected data were: birth weight (in grams) for at-term, singleton deliveries, gestational age (in weeks of amenorrhoea), age of the mother, parity and the outcome of the pregnancy . Individual data for malaria during pregnancy and body dimensions of the mothers were not available.Data on CRFs from the birth registers were entered into a dedicated database written in SQL server 2005. All data were double entered and subsequently exported to MS Access (version 2003). The aggregated monthly malaria data were also imported in the same database. Incomplete records were excluded from analysis and obvious errors  in recorded data were also excluded.For malaria incidence, the total incidence  divided by the population size, was taken for the PHC at which the delivery took place and averaged over the nine months preceding the delivery. This incidence was linked to the individual birth data by PHC and month of delivery. The recorded premature delivery and stillbirth were recorded as obstetric events.Birth weight was analyzed by using the actual numerical value in the statistical models. Low birth weight (LBW) was defined as a birth weight less than 2500 g. The association between malaria incidence and birth weight or obstetric events was first visually inspected by ranking the PHCs according to malaria incidence and plotting birth weight or the rate of obstetric events. Associations between birth weight and other factors and covariates were first explored with standard techniques such as regression and correlation analysis and analysis on variance (ANOVA).The effect of malaria incidence on birth weight was investigated with mixed effects models with birth weight as the dependent variable, and malaria incidence as fixed effect. Other factors and covariates that showed a significant effect on birth weight were subsequently entered in the model. Bayes Information Criterion was used to compare models. The effect of malaria on the rate of obstetric events, stillbirth and abortion, was investigated with logistic regression models following a similar approach.Statistical analysis was done with SPSS . Maps were plotted by Mapinfo .Birth data were collected from 12,526 deliveries in the 11 different PHCs. Incorrect records of 280 deliveries were excluded, leaving 12,246 for analysis, comprising 11,879 at term deliveries, 149 premature deliveries and 218 still births. The data of eligible records are summarized in the Table There was a statistically significant difference in birth weights among health centers  as well as in the frequency of the LWB . However, this difference was mainly caused by the data from Mubuga. There was no explanation pointing at a structural error at this study site. Re-analysing the data without the data from Mubuga yielded similar results. Overall, there was a significant difference of birth weight between the years of study (Anova: p < 0.001). The lowest mean birth weight was recorded in 2003 after which it steadily increased. Figure The mean birth weight for each health center and the mean monthly incidence of confirmed and presumed malaria  is presented in Figure -10, 95% CI: 3.32•10-19 – 0.628; p-value 0.045) and stillbirth . There was also a slight trend towards declining frequencies of stillbirth and pre-mature delivery over the years .Malaria incidence and obstetric events, i.e. pre-mature delivery, and still births, by PHC, are presented in Figure This study confirms that in areas with low to moderate malaria transmission intensity, such as in Rwanda, the complications of malaria in pregnancy are not dominated by low birth weight. Acute symptomatic malaria infections that may cause acute obstetric complications were not reflected in a higher number of premature deliveries and stillbirths in this study.The present research is the first study performed retrospectively on the outcome of pregnancy in the face of a changing pattern of malaria incidence. The strength of the current study is the large sample size that covers a wide geographical range and time interval. The use of stored information also has its limitations. For example, it cannot be explained why the birth weight data from Mubuga are on average higher compared to other PHCs. It cannot be assessed whether a systematic error in recording these data has been made or that indeed the general birth weight is higher in this part of the country due to other, for example nutritional, factors. Data of malaria in the individual pregnant women were not available and the risk of malaria that was used in this study applies to the general population, which can be different from the risk of pregnant women. The pregnant women who deliver at the PHC probably reflect the same population as patients who seek help for symptomatic malaria, but this could not be confirmed in this study.Malaria (in pregnancy) was considered a major health problem in Rwanda  but receThe decline of malaria incidence over the years as reported by National Malaria Control Programme of Rwanda (unpublished data) coincided with the secular trend in birth weight but was not the cause. This shows that, independent from malaria, the conditions of pregnant women are steadily improving. Improved socio-economic factors and consequent improved nutritional status probably play a significant role. In addition, the implementations of vigorous malaria control methods have contributed to the reduction of the disease in Rwanda and this clearly has positive effects on general well-being.The implication of the findings in the present study is that in low to moderate transmission regions such as Rwanda, malaria does not cause LBW. It is mainly a problem of the high transmission regions. The pathophysiology of malaria during pregnancy in holoendemic regions is characterized by chronic or repeated placental infection, inflammation and eventually dysfunction that leads to LBW. In low transmission areas, with low levels of immunity, pregnant women probably have an equal chance as non-pregnant individuals for developing symptoms that will lead to treatment. The disease is thus dominated by a short episode of very symptomatic acute malaria. This has different effects on placental function and fetal conditions than chronic placental infections.The interaction between birth weight and age of the mother (weight increases with mothers' age), which was also found in Tanzania where it was reported that the risk of delivering a baby with LBW was two fold higher in women below 20 years compared to women in age group 20 – 34 years .There was a seasonal relationship in birth weight with an increase in the period June – September, the dry season, of each year. A similar seasonal effect was also found in Tanzania . This peThe present study also reports a negative correlation with birth weight and altitude. In Tanzania, Mmbando and co-workers found a significant decreasing trend of the incidence of LBW from rural lowlands to highlands .There was a slight trend towards declining frequencies of stillbirth and pre-mature delivery over the years. A striking observation is that in Mashesha the highest incidence of malaria was found, but only few cases of premature delivery and no stillbirths occurred. In contrast, the lowest malaria incidence outside Kicukiro , was recorded in Mbuga, where a very average number of stillbirths and pre-mature deliveries were observed.Interestingly, Rwanda had and is still not having specific interventions for pregnant women but takes preventive measures aiming as the population as a whole. Although malaria does not seem to have a major effect on birth weight in Rwanda, pregnant women and their unborn babies are still a very vulnerable group and are at risk for severe effects of malaria episodes during pregnancy. Therefore special attention should be given to this group, also in the absence of low birth weight. In addition trends or fluctuations of malaria in pregnancy should continuously be monitored so that, if needed, specific measures can be taken such as the reintroduction of Intermittent preventive therapy in pregnancy, targeted education for this specific group about risks and preventive measures or targeted bed net distribution.Malaria is not the main determinant of birth weight in Rwanda. Further increase in birth weight could be achieved by nutritional improvements and socio-economic development. Malaria specific interventions remain important and special attention should be given to pregnant women, although they may not increase mean birth weight.The authors declare that they have no competing interests.SR: design and coordination of the study and supervision of data collection and entry, data analysis, drafting of manuscript. PFM: design of the study, interpretation of data, critical reading and modifications of the manuscript, principle investigator. CK: making PNLP malaria data accessible. HDFHS: conception and design of the study, critical reading and modifications of the manuscript. NK: data collection. JV: day-to-day overall supervision of the study. PJdV: conception and design of the study, analysis and interpretation of data, critical reading and modifications of the manuscript, principle investigator


Although behavioural dominance of the right hand in humans is likely to be under genetic control, departures from this population norm, i.e. left- or non-right-handedness, are believed to be influenced by environmental factors. Among many such environmental factors including, for example, low birth weight, testosterone level, and maternal age at birth, season of birth has occasionally been investigated. The overall empirical evidence for the season of birth effect is mixed.We have investigated the effect of season of birth in an epidemiologically robust sample of randomly selected young people (n = 977), all born in the same year. A Kolmogorov-Smirnov type statistical test was used to determine season of birth.Neither the right-handed nor the non-right-handed groups demonstrated birth asymmetry relative to the normal population birth distribution. There was no between-group difference in the seasonal distribution of birth when comparing the right-handed to the non-right-handed groups.The present study failed to provide support for a season of birth effect on atypical lateralisation of handedness in humans. Functional dominance of the right hand is the norm across different populations, various geographical regions, and diverse cultures, with approximately 90% of humans exhibiting clear dominance of the right side of the body. This behavioural characteristic is considered as uniquely human, as there is no other species that displays such a large behavioural asymmetry at the population level. It is also widely accepted that this behavioural feature emerged at some point during the hominid evolution, and that this feature preceded the evolution of another uniquely human feature – language, and in particular, speech as its central component .The transmission of handedness over many generations of humans is widely believed to be under genetic control -4, ratheThe environmental factors believed to provide structural brain substrate for left-handedness include birth difficulties , prenataThe aim of this study is to examine season of birth as a risk factor for hand preference. We used a population-representative random sample of high school students, grouped according to hand preference. We compared their birth distribution with the corresponding age-matched birth distribution in the general population.A total of 1 224 high school students participated in this study. The sample comprised all year 9 students  from six randomly selected high schools in Belgrade, Serbia. Students born in 1989 and 1991 and those with a missing date of birth were excluded from analysis, leaving thus 977 students born in a single year (1990). The reduced sample comprised 457 males (9.4% left-hand writers) and 520 females (5.4% left-hand writers). Whole population birth data were provided by the Statistical Office of the Republic of Serbia.Handedness was assessed by the Edinburgh Handedness Inventory . This inDate of birth was collapsed into a single 12-month frequency series. Season of birth was determined using a Kolmogorov-Smirnov type statistical test . This teNeither the right-handed nor the non-right-handed groups showed a birth asymmetry relative to the normal population birth distribution. Furthermore, there was no between-groups difference in the seasonal distribution of birth comparing the right-handed to the non-right-handed groups. Figure The results of the present study do not support an association between the distribution of handedness and season of birth in young people.per se, have been well described .Although there is some evidence for a season of birth effect on human handedness, the empirical data are inconclusive as there are studies showing contradictory results. For example, several published studies ,25,35,36In conclusion, we found no evidence that season of birth possess an aetiological relevance for developing atypical lateralisation of hand preferences.The author(s) declare that they have no competing interests.The authors all contributed equally to the manuscript, were all involved in the drafting of the manuscript and gave final approval on the manuscript.


In a Norwegian, prospective study we investigated breast cancer risk in relation to age at, and time since, childbirth, and whether the timing of births modified the risk pattern after delivery. A total of 23 890 women of parity 5 or less were diagnosed with breast cancer during follow-up of 1.7 million women at ages 20–74 years. Results, based on Poisson regression analyses of person-years at risk, showed long-term protective effects of the first, as well as subsequent, pregnancies and that these were preceded by a short-term increase in risk. The magnitude and timing of this adverse effect differed somewhat by birth order, maternal age at delivery and birth spacing. No transient increase in risk was seen shortly after a first birth below age 25 years, but an early first birth did not prevent a transient increase in risk after subsequent births. In general, the magnitude of the adverse effect was strongest after pregnancies at age 30 years or older. A wide birth interval was also related to a more pronounced adverse effect. Increasing maternal age at the first and second childbirth was associated with an increase in risk in the long run, whereas no such long-term effect was seen with age at higher order births. From the numerous studies of reproductive history and breast cancer risk, there is a consensus that an early first birth and increasing number of full-term births are associated with a long-term reduction in risk . HoweverThe present study aimed to obtain detailed information on associations between breast cancer risk and timing of births based on follow-up information for 1 691 555 Norwegian women of parity 0–5, including 22 890 breast cancer cases at ages 20–74 years.The present study includes all Norwegian women born in the period 1925 to 1979, who had been residents of Norway for some period after 1960 and thus were included in the Norwegian Population Registry. The present data set is an update (until 1 January 2000) and extension of our previous data set  with inf5 person-years in the age interval 20–74 years. The mean follow-up time was 24.9 years, range 1 month to 54 years.Each woman was considered at risk of breast cancer from age 20 years. However, since the Cancer Registry of Norway was established in 1953, no women entered the risk set before 1955. Thus, women born in 1925–1934 entered the risk set at age 21–30 years. At closing date of study, the oldest women were 74 years old. A total of 1 691 555 women of parity 0–5 were included in follow-up, contributing a total of 3788 × 10Only cases classified as primary malignant neoplasms of the breast  were considered. A total of 22 890 women were diagnosed with breast cancer during follow-up, supported by histological examination and/or by autopsy for 22 666 (99%). Of these, 22 510 (99.3%) were classified as carcinomas, 107 (0.5%) as sarcomas and 49 (0.2%) as other or unspecified tumours.2 and (age)3, which gave a very good fit to the observed age-specific incidence rates among nulliparous women. Stratification was made on birth cohort (11 five-year categories).Associations between breast cancer incidence and the reproductive factors were examined in a log-linear Poisson regression model of person-years at risk . To circParity was treated as a time-dependent categorical variable, represented by indicator variables in the model. A woman entered a new exposure category (higher parity group) whenever she gave birth to a child. Only women of parity 5 or less were considered. Nulliparous women were assigned a constant value of age at births (corresponding to the chosen reference groups) and time since birth  to make it possible to estimate the age effect on the basis of this group , or as a categorical variable (5-year groups). Among women with five full-term births, only 5-year categories were considered .Time-dependent variables were defined also for age at each childbirth, with a constant reference value assigned until date of birth, and actual age at date of delivery otherwise. In this way, the terms for age at A restricted cubic spline curve  was usedIn joint analyses of age at, and time since, a particular childbirth, adjustment was made for age at all previous births. Owing to collinearity, the estimated associations with age at previous births will partly reflect relations with time since previous births. Interaction terms were examined whether the association with time since birth, as represented by the spline regression coefficients, differed by order of birth, age at delivery or birth interval. Based on the values of the estimated regression coefficients, combined with coefficients for parity and age at births, the difference in estimated log-rate between a particular subgroup of parous women and nulliparous women over time was calculated. An antilogarithmic transformation was made to determine incidence rate ratios. For particular purposes, the incidence rate according to attained age was calculated, with time-dependent contribution of the reproductive factors.Tabulation of individual records into person-years tables and Poisson regression analyses of person-years at risk, with calculation of maximum likelihood estimates and likelihood ratio tests, were performed by means of the EPICURE program package  are shown in vs uniparous women were 0.91 (0.85–0.96), 0.81 (0.73–0.89), 0.64 (0.57–0.72), and 0.50 (0.38–0.66), respectively. However, the association with time since birth differed by birth order  for women with two to five full-term births P=0.58). Based on the interaction model .The youngest biparous mothers (second birth <25 years) experienced a transient increase in risk shortly after birth, whereas no adverse effect appeared among women with a second birth at age 25–29 years . Among tP=0.055). In general, a narrow birth interval (<3 years) seemed to result in a high initial risk followed by a delayed adverse effect, whereas a wide birth interval was associated with a faster and more long-lasting elevation in risk.Additional subgroup analyses showed that the sharp transient increase in risk among the youngest mothers was restricted to those with first and second birth close in time . A transient increase in risk was seen among women with a third birth at age ⩾30 years only, but with an earlier peak in risk among the oldest mothers (⩾35 years). Age at the third birth seemed to be of less importance for the long-term risk, as the risk curves in the different subgroups appeared to overlap after some time.The risk pattern after third birth differed significantly by the mother's age at delivery , P=0.026Further analyses according to birth spacing showed a transient increase in risk also after a third birth at a younger age (<30 years), unless the second birth had occurred within the last 3 years . Among tvs B, 5D vs E and 5C vs F). The transient increase in risk after a third birth was particularly strong among triparous women who had their first birth at age 25–29 years, with a wide birth spacing between the first and the two subsequent births  had a higher risk than nulliparous women until about 27 years after delivery, that is, at an age of at least 57 years. With a first birth at a lower age, the risk curves crossed earlier. In terms of absolute rate, the predicted crossover point was somewhat earlier due to some variation in the age estimates between the different analyses. Based on the ‘breast-tissue age’ model , but theWe found an association with age at the second birth among biparous women with long-term risk differing by second birth below and above age 25 years. The few studies that found an association with age at second birth (Consistent with results of previous studies , we obseAn association between mitotic activity of the breast tissue and hormonal factors, in particular those related to oestrogens and prolactin levels, but possibly also progesterone, was a main concept underlying the breast-tissue age model of In summary, the present study showed a rather complex association between reproductive history and breast cancer risk. The mother's age at births, as well as birth spacing, influenced the magnitude and timing of the transient increase in risk shortly after a delivery. In the long-term, however, only the number of births and the mother's age at the first and second birth seemed to be important for the risk level. The changes in childbearing pattern during recent decades, with fewer children and higher age at births (


Anogenital distance (AGD) is sexually dimorphic in rodents and humans, being 2- to 2.5-fold greater in males. It is a reliable marker of androgen and antiandrogen effects in rodent reproductive toxicologic studies. Data on AGD in humans are sparse, with no longitudinal data collected during infancy.This study was designed to determine AGD from birth to 2 years in males and females and relate this to other anthropometric measures.Infants were recruited from the Cambridge Baby Growth Study. AGD was measured from the center of the anus to the base of the scrotum in males and to the posterior fourchette in females. Measurements were performed at birth and at 3, 12, 18, and 24 months of age.p < 0.0001). AGD increased up to 12 months in both sexes and in a sex-dimorphic pattern. AGD was positively correlated with penile length at birth  and the increase in AGD from birth to 3 months was correlated with penile growth .Data included 2,168 longitudinal AGD measurements from 463 male and 426 female full-term infants (median = 2 measurements per infant). Mean AGD (± SD) at birth was 19.8 ± 6.1 mm in males and 9.1 ± 2.8 mm in females (We report novel, longitudinal data for AGD during infancy in a large U.K. birth cohort. AGD was sex dimorphic at all ages studied. The availability of normative data provides a means of utilizing this biological marker of androgen action in population studies of the effects of environmental chemicals on genital development. The trends in male reproductive health characterized by increases in the incidence of cryptorchidism, hypospadias, testicular tumors, and declining semen quality observed in many countries have been the subject of a number of recent reviews . Geograpin utero exposure to androgens and chemicals with antiandrogen effects . It is measured from the center of anus to the genital tubercle and is sexually dimorphic, being 2- to 2.5-fold longer in males. Numerous studies attest to its validity as a sensitive marker for the effects of  effects . Consequ effects . Neverth effects . It has  effects . The aimInfants were participants of the Cambridge Baby Growth Study, an ongoing prospective, longitudinal study established in 2001 to characterize hormonal, genetic, and environmental influences on infant growth and early male reproductive development. Measurement of AGD was included in the study protocol from 2006. Mothers were recruited from the maternity unit of Addenbrooke’s Hospital, Cambridge (UK), at around 12 weeks of gestation when attending for their routine antenatal ultrasound scan. The unit conducts about 5,000 deliveries per year and supports a mixed urban and rural community in South Cambridgeshire. Information about maternal lifestyle was collected using self-administered questionnaires. Detailed analysis of this information is the subject of a separate study. The research protocol was approved by the Cambridge Local Research Ethics Committee, and the study was conducted in accordance with the International Conference on Harmonization standards for Good Clinical Practice. The mothers gave written informed consent for themselves and their infants to participate in the study before the recruitment. The infants were measured at birth either at the hospital or at home and subsequently during infancy in a dedicated follow-up research clinic by a team of four research nurses.Infants were measured at birth and at 3, 12, 18, and 24 months of age. Measurements of AGD, weight, length, and penile length were collected, and the external genitalia were examined for any abnormalities.A modified version of a previously published method by The research nurses received regular training within the department and participated in annual quality control exercises when an anthropometrist supervised their measurement techniques. The interobserver variability was assessed by repeated measurements in 14 infants, and the technical error of measurement (TEM) was calculated . The absp-Values of < 0.05 were considered statistically significant. Centile curves were created using lmsChart-Maker Light version 2.3 . Results are presented as mean (± SD) except as specified.The gestational age was calculated from the last menstrual period. Preterm (gestation < 37 weeks) and low-birth-weight  infants were excluded from analysis. Subjects whose measurements were performed beyond a 4-week interval of the proposed date of follow-up were also excluded. Differences between variables were determined by using analysis of variance . We used Pearson’s correlation coefficient to estimate the strength of associations. We assessed the longitudinal change in AGD by calculating the difference between neighboring time points in infants who were measured on both occasions. AGD and other variables were assessed by multiple linear regression models using weight, length, penile length, gestational age, and age at examination as independent variables. We assessed AGD in 889 infants . Twenty infants were born of twin gestation. Birth weight of infants were similar to 1990 British reference data, in the form of mean standard deviation score [SDS (± SD)]: male, 0.08 ± 0.88; female, 0.14 ± 0.88 . Howeverp < 0.0001)  measurements per infant] . Males h 0.0001) . At birt 0.0001) . AGD at  0.0001) .r = 0.16, p = 0.006). In females, AGD was correlated with weight at birth, 3, 12, and 18 months and with length at 18 and 24 months of age. When adjusted for other variables in a linear regression model using weight, length, gestational age, and age as independent variables the positive association persisted only for length at 24 months of age .At each time point, we analyzed AGD for correlation with respective measures of body size. In males, AGD was correlated with weight at 3 months and with length at 3 and 12 months of age . In a lir = 0.18, p = 0.003). We also observed a correlation between the increase in AGD between birth and 3 months of age and an increase in penile length during the same period . Neither of these parameters showed any correlation with the increase in weight or length during the same period. In a linear regression model using weight, length, gestational age, penile length, and age as variables, birth weight and penile length were the only predictors of AGD at birth . Penile growth was the only predictor for the increase in AGD during the first 3 months of life  in a linear regression model using increase in weight and length as independent variables. Similarly, in females increases in weight and length from birth to 3 months of age were not correlated with an increase in AGD during the same period.AGD was correlated with penile length only at birth . In a preliminary analysis, the mean AGD in 13 infants with incomplete testicular descent at birth was shorter than in males with normal testis descent (17.5 mm vs. 19.9 mm) . HoweverThis study confirms sexual dimorphism of AGD at birth that was maintained to the same degree during the first 2 years of life. In males, AGD was related to penile length and penile growth and hence may be useful as a marker of androgen action.Perineal growth and caudal migration of the genital tubercle in male rodents are androgen (dihydrotestosterone)-dependent processes . ConsequThe mean AGD in males at birth in the study (19.8 mm) is similar to the measurements published by AGD rapidly increased from birth to 12 months of age and then gradually plateaued, resulting in a 70% increase in this measurement during the first 2 years of life in both sexes. The reported length of the perineum measured from the anterior anal margin to the posterior fourchette in adult females (31.3 mm ± 8.5) is almost double the mean AGD at 2 years of age . Even thIn rodents, birth weight is not a strong predictor of AGD . In the The degree of sexual dimorphism in AGD was consistent at all stages of the 2-year longitudinal study. The association of AGD with the penile length at birth and penile growth in the first 3 months may reflect the effect of prenatal androgens and the postnatal surge in testosterone secretion, respectively. The reported strong association of penile growth in early infancy with serum testosterone levels at 3 months of age  providesp-chlorophenyl)ethylene (DDE), a metabolite of the pesticide 1,1,1-trichloro-2,2-bis(p-chlorophenyl)ethane (DDT)  was reported in three female infants with virilizing congenital adrenal hyperplasia . A decrene (DDT) . In contne (DDT) .Measurement of AGD offers a potentially more sensitive marker of the effects of prenatal exposure to EDCs in humans compared with possible indices of endocrine-disrupting effects such as cryptorchidism and hypospadias. The longitudinal normative data from this study is fundamental to applying the technique of AGD measurement for future cross-sectional population studies of the effects of EDCs in human populations.The present population-based prospective study reports for the first time normative data for AGD in both sexes from birth to 2 years of age. We observed a sex-dimorphic pattern at all ages studied. Growth of the perineum as determined by measurement of the AGD was characterized by a rapid increase in the first 12 months of life, followed by gradual plateauing in both sexes. The association with penile length indicates that AGD may be influenced by both prenatal and early postnatal androgen secretion. By the same token, any antiandrogen effects during this period of human development may be reflected in decrements in AGD measurements. The availability of longitudinal normative data for AGD now provides a means to determine the reliability of this marker as a sensitive indicator of the effects of EDCs in human populations.