Birth weight and length have seasonal fluctuations. Previous analyses of birth weight by latitude effects identified seemingly contradictory results, showing both 6 and 12 monthly periodicities in weight. The aims of this paper are twofold: (a) to explore seasonal patterns in a large, Danish Medical Birth Register, and (b) to explore models based on seasonal exposures and a non-linear exposure-risk relationship.Birth weight and birth lengths on over 1.5 million Danish singleton, live births were examined for seasonality. We modelled seasonal patterns based on linear, U- and J-shaped exposure-risk relationships. We then added an extra layer of complexity by modelling weighted population-based exposure patterns.The Danish data showed clear seasonal fluctuations for both birth weight and birth length. A bimodal model best fits the data, however the amplitude of the 6 and 12 month peaks changed over time. In the modelling exercises, U- and J-shaped exposure-risk relationships generate time series with both 6 and 12 month periodicities. Changing the weightings of the population exposure risks result in unexpected properties. A J-shaped exposure-risk relationship with a diminishing population exposure over time fitted the observed seasonal pattern in the Danish birth weight data.In keeping with many other studies, Danish birth anthropometric data show complex and shifting seasonal patterns. We speculate that annual periodicities with non-linear exposure-risk models may underlie these findings. Understanding the nature of seasonal fluctuations can help generate candidate exposures. Birth weight has long been acknowledged as an important measure of neonatal health . In addiApart from seasonal fluctuations in birth weight, birth length also has an annual periodicity ,9,10. UsOur ability to analyse seasonality in time series data has improved substantially in recent decades, with more robust and flexible statistical models now available to capture the complexities of the data . RecentlThe interpretation of these data has been a challenge to the research community. The presence of both 6 month and 12 month signals have lead to the understandable conclusion that at least two different 'source generators' must be driving these signals. To further complicate matters there are two scenarios, either: (a) two source generators of different amplitudes, each with 12 month periodicities, but phase shifted by 6 months (summing to produce both 6 and 12 month periodicities), or (b) one source generator with a signal of 12 months and a second source generator with a nested harmonic with a 6 month period. It is difficult to identify biologically plausible candidate exposures that would explain such patterns.To date, seasonality models have been based on the unspoken assumption that the Exposure-Risk Relationship (ERR) is linear. In other words, as the seasonally-fluctuating exposure changes over the year, the biological variable of interest responds in a linear fashion. However, J-shaped or U-shaped ERRs are relatively common in perinatal epidemiology and related research -22. It sA second unspoken assumption of seasonal analyses to date is that the entire population experiences the full range of exposure. There is reason to believe that some portions of the population may only experience parts of the exposure range. For example, some portion of the population may be 'partitioned' to one segment of the J- or U-shaped ERR. Using our previous hypothetical relationship between ambient temperature during the third trimester and birth weight, if sites that were exposed to only one temperature extreme , then seasonal analyses from such sites would show a 12 monthly periodicity. By partitioning different fractions of the populations within different subregions of the J- or U-shaped ERR (e.g. the linear upper segment versus the nonlinear lower segment), we predict that the resultant summed time series of the dependent variable will be altered.We are not aware that these two assumptions have been previously explored in seasonality research. Because the models involve a number of stages and are perhaps not intuitive, modelling exercises can assist in demonstrating their properties. We aim to provide alternative models that could help explain some of the complexities of seasonal fluctuations in birth weight and birth length data, as well as explain some of the differences in published results.In the first part of the results we will present the data from the Danish Medical Birth register , and in Data were derived from the Danish National Centre for Register-based Research. We examined birth weight and birth length in all live-born, singletons with gestations of at least 37 weeks born between 1973 and 2002 (includes births in 2002). In order to reduce the possible influence of migrant-status on birth anthropometry, we included only births where both the mother and father had been born in Denmark.t) is modelled using a sinusoidal form:We applied similar techniques to those used in the study of Australian perinatal registers . Recordswhere the phase controls the location of the seasonal peak, and the amplitude controls its size. For the analyses presented here the period was fixed to follow a 6 or 12 month cycle. It is possible to specify multiple seasonal components with different periods. An important feature of this method is that it is a dynamic model, and hence the estimated seasonal pattern is allowed to change over time. Methods based on averaging the entire dataset by month ,24 assumt is dependent on the estimate at time t-1. Using this structure any change in seasonality (e.g. a decreasing amplitude over time) is assumed to occur gradually.This is known as a non-stationary model, because season is not necessarily fixed in time. The non-stationarity in the amplitude and phase are modelled using an autoregressive structure, so that the seasonal estimate at time Because the model is dynamic and involves the non-linear cosine function it cannot be fitted using standard methods, such as maximum likelihood. Instead the method uses Markov chain Monte Carlo (MCMC) to estimate the joint distribution of the parameters. These distributions are estimated empirically by iteratively sampling each parameter. Given the empirical joint distribution the mean and 95% confidence interval are then easily calculated. We used 5000 MCMC samples with a burn-in of 1000.The model assumes that the residual noise is normally distributed, and this assumption was checked using the Shapiro-Wilk test. Because the original paper by Wohlfahrt and colleagues  reportedWe combined data for males and female babies. Although males are generally heavier than females at birth, the trends over time and seasonal changes in weights and lengths were similar. Hence combining the data for this analysis is justified. We plotted the estimated trends  in birth weight and length over time. We also plotted estimates of the overall seasonal pattern by time. Because the time series is long, we plotted seasonal patterns once every five years, and also for the last year (2002).In order to explore the assumptions of standard seasonal analyses, we first modelled time series data with a seasonal (i.e. 12 month) periodicity, but varied both the nature of the exposure-risk relationship , and the distribution of the population within the exposure range. The exposure-risk relationship is analogous to a dose-response relationship. The outcome may not always be a risk (e.g. odds of disease), but may be a continuous measure (e.g. birth weight).If different populations are exposed to limited doses of the exposures (e.g. as might be expected at different latitudes for temperature and ultra-violet radiation-related exposures), then the resulting time series may also reflect the 'local' properties of the non-linear curve (i.e. the population maps on to a limited section of the J-shaped curve). We assumed a single symmetric sinusoidal seasonal exposure (a period of 12 months), with a peak in January. We examined a range of likely exposure-risk relationships and generated the associated time series of seasonal risk. The exposure-risk and seasonal-risk relationships were plotted. In a second analysis we varied the Population Exposure Distribution (PED) into low, medium and high groups. We assigned a number of different weights to these groups to simulate different populations.Finally, by way of demonstration, we attempted to build a model that mimics key aspects of the Danish perinatal data. We assumed a fixed J-shaped ERR across time, and examined the impact of shifting the PED on the seasonal distribution.The Danish Medical Birth Register included 1 573 203 neonates . Over the entire epoch, the mean (and standard deviation) for birth weight in males and females was 3533.2 (578.9) and 3412.0 (545.8) grams respectively. For birth length, the mean (and standard deviation) for males and females was 52.2 (2.6) and 52.4 (2.5) centimetres respectively.Figures Figures Concerning the first modelling exercise, Figure Changing the ERR from linear to U-shaped generated a time series with a 6 month periodicity (i.e. double the frequency of the parent 12 month periodicity). Altering the ERR to J-shaped partitioned the resulting time series in a dominant 12 month signal and a secondary 12 month signal phase-shifted by 6 months.The seasonal patterns in Figure Finally, Figure In keeping with a previous study also based on the Danish population sample , we founIn general, we note that J-shaped or U-shaped models of exposure-risk result in substantial changes in the nature of seasonally-generated time series. The findings have implications for those wishing to synthesise or understand diverse and apparently conflicting seasonal data. Time series with both 6 and 12 month periodicities may be the result of an exposure with a J-shaped exposure-risk relationship.Furthermore, if researchers are not aware of the underlying U- or J-shaped models, different populations facing different doses of the exposure could result in conflicting results. For example, as shown in Figure The results of this study highlight that interpreting overall seasonal patterns should be done with care. The simplest interpretation requires the assumption that exposure is constant across the population. Many researchers examining seasonal patterns may not realise that they are making this assumption, or the implications if it is violated. This could lead to them wrongly concluding that two seasonal exposures are at work (e.g. a 6 and 12 month exposure).Our analyses of Danish birth anthropometric data clearly show that the amplitudes did not remain constant over time Figure . This meDepending on the cancellation of seasonal components  declare that they have no competing interests.JM and AB designed the study. AB performed the analyses. PBM and CBP provided the birth weight data. DE and TB contributed to the interpretation of the data. All authors contributed to the interpretation of the data and writing up, and all authors read and approved the final document.The pre-publication history for this paper can be accessed here:


This study investigates how secondhand smoke (SHS) exposure influences neonatal birth weight in Jordan, a country with high smoking prevalence. The findings revealed that as the average number of SHS exposure hours per week increased in the second trimester, the neonatal birth weight decreased while holding all covariates constant. Women who reported a higher average number of SHS exposure hours per week from work in the second trimester, home in the third trimester, and outside in the third trimester were at greater risk for having a low birth weight neonate than women who reported a lower average number of SHS exposure hours. This may be of particular importance in Jordan due to the high prevalence of SHS exposure experienced by non smoking women. Although SHS is a known cause of premature death and disease among women and children , most st1.2.In this prospective survey study, the influence of SHS exposure on neonatal birth weight and the risk of having a LBW neonate as a result of Jordanian non-smoking women pregnant women being exposed to SHS were investigated. The research hypotheses were:Increasing the reported average number of SHS exposure hours per week from home, work, and outside in the first, second, and third trimesters lead to decreasing the neonatal birth weight significantly after controlling for covariates.Women who reported higher average number of SHS exposure hours per week from home, work, and outside in the first, second, and third trimesters were at significantly greater risk for having a LBW neonate than women who reported lower average number of SHS exposure hours after controlling for covariates.Among non-smoking Jordanian women giving birth at a major medical center in northern Jordan:1.3.Jordan is a developing, low-to-middle income country that is advancing and adapting to a twenty-first century economy. It is geographically situated between Saudi Arabia, Iraq, Israel, and Syria and thus is directly embedded in the culture, societal trends, and concerns of the Middle East as a whole. Administratively, Jordan has three regions with 12 governorates. The gross domestic product is $8.9 billion, which yields a mean of $1,744 per capita .Jordanians view smoking as a social habit; having coffee and cigarettes with friends and family members is deeply rooted in the culture. From 2002 to 2006, 50% of adult men and 11% of adult women smoked –14. A reJordan and many other Middle Eastern countries lack systematic knowledge and surveillance data about the harm and economic effects of smoking and second-hand smoke exposure. Unfortunately, many aspects and social attitudes of using tobacco are shaped by their acceptance in Jordanian Arab culture . Therefo1.4.Few reports chronicling the possible influence of SHS exposure among non-smoking Jordanian pregnant women on birth weight in this population were found. Further, LBW neonates have a greater risk for mortality and morbidity in the first week of life . TherefoSecondhand smoke exposure. Pregnant women are usually exposed to SHS in different places for different durations and at different times during pregnancy. For the purpose of this study, the assessment of SHS exposure included three major components: the type of SHS exposure , the quantity of SHS exposure (average number of exposure hours per week), and the timing of the SHS exposure .Direct effect of SHS exposure. The toxins in SHS directly cause harmful effects in the fetus [he fetus –28. Nicohe fetus –34. Cotihe fetus ,35–37.Birth outcome. For the purpose of this study, birth outcomes included low birth weight and actual birth weight. Birth weight was defined as the number of grams that represented the neonate’s weight at birth. Low birth weight was defined as a birth weight of less than 2,500 grams1.5.The rate of newborn, infant, and early childhood growth is a sensitive indicator on the impact of SHS exposure. Multiple maternal and neonatal factors are also associated with birth weight. Two separate secondary data analyses of survey data for 35 developing countries showed that maternal education and household wealth have a strong positive association with height for age and early life diseases, with substantial shared variance between these two variables in explaining child height for age ,39. Clea2.2.1.Princess Badia Teaching Hospital in Irbid, northern region of Jordan served as the setting for recruiting study participants for this study. This hospital is part of the Ministry of health (MOH) Public health sector hospitals in Jordan and is the main public hospital that specializes in obstetric and gynecology in the northern region of Jordan. About 900 deliveries occur in this hospital every month. In Jordan, 95% of the deliveries occur in hospitals. 65% of these deliveries occur in public MOH hospitals .2.2.Given a power analysis using a moderate effect size of 0.50 and a priori power estimation of 0.80, for a two-tailed test with alpha previously set at 0.05, at least 67 subjects for each group are required to have confidence in the findings . Birth wSystematic sampling was used for this study. On each day that was designated for data collection, the current patient list on the post delivery floor was reviewed. Every third patient was selected as a potential participant prior to knowing about all the inclusion criteria. Participants met eligibility requirements if they were 18 years of age or older at the time of recruitment, reported that they did not smoke tobacco for one year prior to conception, their pregnancies resulted in a single live birth, and their pregnancies resulted in full term births . The eligibility criteria of having a non-malformed full term birth were also verified from the medical records. Women who were not aware of having malformed or preterm births were excluded after they were asked to fill out the SHS exposure questionnaire.Mothers who had medical problems diagnosed before pregnancy , or medical problems diagnosed during pregnancy  were excluded from the study.2.3.The Environmental Tobacco Smoke Exposure questionnaire was used to measure SHS exposure from home, work, and outside in the first, second, and third trimesters. The questions in the SHS questionnaire were adapted from the Pregnancy Risk Assessment Monitoring System (PRAMS) questionnaire . The PRAEnvironmental tobacco smoke exposure at home was defined as the average number of hours per week in which the pregnant women was exposed to SHS from her husband or anyone else inside her house. This information combined with the mother’s report of her time spent in the home was used to help quantify the predictor variable, the average of hours per day the woman was exposed to cigarette smoking by her husband or anyone else inside her house in each trimester.Environmental tobacco smoke exposure at work was defined as the average number of hours per week in which the pregnant woman was exposed to SHS at her job. Environmental tobacco smoke exposure at outside places was defined as the average number of hours per week in which the pregnant woman was exposed to SHS from outside home or work, such as relatives’ homes, restaurants, cars, and doctors’ offices.The SHS exposure questionnaire was also used to measure potential covariates. The potential covariates included neonate’s characteristics, maternal demographic data, maternal obstetric history, and current delivery. The neonate’s characteristics included gender and gestational age. Maternal demographic data included mother’s age, height, total family income, education, and occupation during pregnancy. Maternal obstetric history and current delivery included parity, weight before pregnancy, weight gain during pregnancy, initiation of prenatal care, alcohol use during pregnancy, and illicit drug use during pregnancy.The Arabic version of the SHS Exposure Questionnaire was pilot tested on a group of Jordanian women (n = 30) in order to test data collection procedures and compute the reliability coefficient and construct validity of the Arabic version. The total scale yielded a Cronbach’s alpha of 0.78. A factor analysis using principle components extraction was performed to ascertain whether the Arabic version of the SHS exposure questionnaire items would load in a pattern similar to the one in the English data sets. The pilot study revealed three factors which matched with the original English questionnaire. Together the three factors explained 98.9% of the total SHS exposure variance.Finally, based on the feedback from participants, the last three items of the questionnaire were modified to include outside places of exposure such as relative homes in addition to public places such as restaurants, cars, and doctors’ offices.Medical records were also used to validate self-reported data related to exclusion criteria, including medical problems diagnosed before pregnancy  and medical problems diagnosed during pregnancy . Anemia was defined as hemoglobin concentration of less than 11 gm/dL in the first and third trimester or hemoglobin concentration of less than 10.5 gm/dL in the second trimester . Data reBirth weight was defined as the number of grams that represented neonate’s weight at birth as recorded in the neonate’s medical records. Low birth weight was defined as a neonate with less than 2,500 grams at birth as recorded in the neonate’s medical records. The actual birth weight in grams was obtained from the medical records. Based on birth weight, neonates were classified as LBW or normal birth weight neonates.Finally, data were collected from medical records to measure potential neonatal covariates. These covariates included neonate’s gender  and gestational age . Data requested from the neonate’s record also included the date of delivery and the method of delivery.The committee for protection of human subjects at Jordan University of Science and Technology approved the study design and consenting methods prior to beginning the study. Also, formal permission to conduct this study in Jordan was obtained from the director of Princess Badia Teaching Hospital. A brief description of the study purpose was given to participants before the questionnaire was distributed and voluntary consent was obtained. Participants were informed that their responses would be kept confidential and that all results would be presented as an aggregate data. Participants were asked for their permission to access some information from their medical records and their neonates’ medical records.The unit of measurement for initiation of prenatal care was the month of pregnancy and that in the multiple regressions analysis it was quantified by months. The mother’s occupation during pregnancy was quantified as; the mother was asked if she have a job outside home , if yes, how many weeks did she work in each trimester, how many hours did she work on average per week, While pregnant, if she had regularly been in the same room with someone who smokes cigarettes, and finally on average how many hours per week she has been in the same room with someone who smokes cigarettes at her job in the 1st, 2nd, and 3rd trimester. The last variable was entered in the regression analysis as a predictor of SHS exposure at work.The covariates included in the analysis were neonate’s gender, gestational age, number of mother’s previous pregnancies, weight gain during pregnancy, initiation of prenatal care, maternal age, maternal weight before pregnancy, mother’s years of education, mother’s occupation during pregnancy, and total family income. Mean number of living children included the neonate and all other living, biological children of the participant.3.3.1.The mean age of the study participants was 28 years (SD = 5.52), the mean height was 160.2 centimeters (SD = 6.25), the mean years of education was 11.4 years (SD = 2.79), and the mean total family income was 2,888 $US per month (SD = 105.22). 42.3% of the study participants reported having an education level of less than high school; 33.3% completed high school education, and 24.3% completed post high school education. With regard to the father’s education level, 45% reported having an education level of less than high school; 32.3% completed high school education, and 22.6% completed post high school education. Only 10% of the study participants reported having jobs outside home (n = 31). Many of the demographics fit the national profile as the mean height for women is 160 Centimeters, and 55% of women have more than high school education. However the reported total family income was lower than the national average  [nth SD = 5.22. 42.3.2.The mean weight before pregnancy of the study participants was 63 kilograms (SD = 10.53); the mean weight gain during pregnancy was 11.1 kilograms (SD = 3.48); the mean number of pregnancies was 3.5 pregnancies (SD = 2.24); and the mean number of living children prior to the subject neonate was three ; 76% had vaginal deliveries (n = 228); 24% had caesarean sections (n = 72); and 9.3% had LBW neonates (n = 28).The mean birth weight of the participants’ neonates was 3,245.95 grams (SD = 444.06), and the mean gestational age was 39.3 weeks (SD = 1.19). Furthermore, 50.3% of the participants had baby boys  Furthermore, the average reported number of cigarettes that were smoked by the exposed mother’s husband in the home was 16.3 cigarettes per day (SD = 15.89).3.4.n = 272) and the group of participants who had LBW neonates (n = 28) in terms of potential covariates. Descriptive statistics, t-tests for continuous variables, and chi-square tests for categorical variables were performed to compare the two groups. There were 272 participants who had normal birth weight neonates (90.7%) and 28 participants who had LBW neonates (9.3%). This is similar to the LBW rate in Jordan as a whole (10.6%) [Before running the multiple regression and logistic regression, an equivalent analysis was conducted between the group of participants who had normal birth weight neonates ( (10.6%) . The meat (298) = 3.2, p < 0.001, weight gain during pregnancy t (298) = 2.1, p = 0.036, and gestational age of the neonate t (298) = 2.8, p < 0.005 were significant covariates (p < 0.05) ( < 0.05) . These c2  = 0.69, p = 0.41 (two-tailed), mother having a job outside home χ2  = 1.89, p = 0.17 (two-tailed), or initiation of prenatal care in the first trimester χ2  = 1.87, p = 0.17 (two-tailed). There were no participants who used drugs or alcohol during pregnancy.There was no significant difference between the two groups in terms of neonates’ gender χ3.5.A stepwise multiple regression analysis was conducted to test the first hypothesis, where the dependent variable, which is the neonatal birth weight, could be explained by the independent variable, which is the reported average number of SHS exposure hours per week, in the first, second, and third trimesters. Multicollinearity was examined between the independent variables before the stepwise multiple regression analysis, to test for collinearity. Specifically the highest correlation coefficient between any independent predictor variables was only 0.49. The variance inflation for the independent variables was calculated with a result of 0.91, which indicated a high level of tolerance thus low collinearity.The potential covariates that were entered in the model included the neonate’s gender, gestational age, number of mother’s previous pregnancies, weight gain during pregnancy, initiation of prenatal care, maternal age, maternal height, maternal weight before pregnancy, mother’s years of education, mother’s occupation during pregnancy, and total family income. From previous literature ,27,38 th2 = 0.38; F  = 30.13; p < 0.05)  .The total neonatal birth weight variance that was explained by the significant independent variables and significant covariates was 38%. The reported average number of SHS exposure hours per week from home, work, and outside in the second trimester explained 22% of the neonatal birth weight variance. The gestational age of the neonate, mother’s weight before pregnancy, and mother’s weight gain during pregnancy explained the remaining 16% of the neonatal birth weight variance.3.6.A logistic regression was performed to test the second hypothesis. The estimated coefficients were the actual measures of the changes in the log odds ratio (ratio of the probabilities) of having LBW as the average number of SHS exposure increased one hour per week in each trimester while holding all covariates constant.Based on logistic regression analysis, women who reported a higher average number of SHS exposure hours per week from occupational exposure in the second trimester were at significantly greater risk for having a LBW neonate than women who reported a lower average number of hours of SHS exposure from occupational exposure after controlling for mother’s weight before pregnancy, mother’s height, mother’s weight gain during pregnancy, total family income, and gestational age of the neonate. The adjusted odds ratio was OR = 1.331 (95% CI 1.052–1.684) for one unit increase of SHS exposure from work in the second trimester (p < 0.05). Women who reported a higher average number of SHS exposure hours per week from home and outside in the third trimester were at significantly greater risk for having a LBW neonate than women who reported a lower average number of hours of SHS exposure after controlling for the same covariates. The adjusted odds ratios were OR = 1.075 (95% C1 1.029–1.124) for one unit increase of SHS exposure from home in the third trimester and OR = 1.154 (95% CI 1.055–1.262) for one unit increase of SHS exposure from outside in the third trimester (p < 0.05) .With regard to significant covariates, subjects who reported a higher average weight before pregnancy were at significantly lower risk of having a LBW neonate than subjects who reported a lower average weight before pregnancy . The adjusted odds ratio was OR = 0.934 (95% CI 0.881–0.992) for one unit increase of mother’s weight before pregnancy. Subjects who reported a higher average weight gain during pregnancy were at significantly lower risk of having a LBW neonate than those who reported a lower average weight gain during pregnancy . The adjusted odds ratio was OR = 0.856 (95% CI 0.739–0.991) for one unit increase of weight gain.2 = 0.38; F  = 30.13; p < 0.05). The SHS exposure variables accounted for 22% of the birth weight variance. Also, subjects who reported higher average number of SHS exposure hours per week from occupational exposure in the second trimester and from home and outside in the third trimester were at significantly greater risk for having a LBW neonate than women who reported lower average number of SHS exposure hours (p < 0.05). The adjusted odds ratio were OR = 1.331 (95% CI 1.052–1.684) for work exposure, OR = 1.075 (95% C1 1.029–1.124) for home exposure, and OR = 1.154 (95% CI 1.055–1.262) for outside exposure.In summary, as the reported average number of SHS exposure hours per week from all places  in the second trimester increased, the neonatal birth weight significantly decreased while holding all covariates constant  significantly reduced birth weight. The last two trimesters were the times in which SHS exposure significantly reduced birth weight. The findings related to timing of exposure were of particular importance since it was not investigated in the previously reviewed studies.The study results are consistent with the conclusions of the World Health Organization. The International Consultation on Environmental Tobacco Smoke and Child Health concluded that maternal smoking during pregnancy is a major cause of reduced birth weight, decreased lung function, and sudden infant death syndrome (SIDS). Exposure to SHS among nonsmoking pregnant women can also cause a decrease in birth weight . HoweverSmoking is on the rise in developing countries. Environmental conditions such as overcrowding may make the health effects of SHS more pronounced . HoweverIn summary, based on the results of this study, SHS exposure during pregnancy exerts significant adverse effects on birth outcome. These results are consistent with the studies that were conducted in the developed countries, in the developing countries, and with the recommendations and conclusions of the WHO. The findings of this study add new support to the growing evidence showing adverse effects of SHS exposure on birth outcome. The quantity, type, and timing of exposure were all significant predictors of birth weight.4.1.In this study, mother’s weight before pregnancy, mother’s weight gain, and gestational age of the neonate were significant predictors of neonate’s birth weight and accounted for 16% of its variance. Furthermore, mother’s weight before pregnancy and weight gain during pregnancy significantly decreased the risk of having a LBW neonate. These results are consistent with the results of previous studies. Higher weight gain during pregnancy and higher maternal pre-pregnancy weight were found to significantly decrease the risk of having a LBW neonate . MaternaIn summary, mother’s weight before pregnancy, mother’s weight gain, and gestational age of the neonate were significant predictors of neonate’s birth weight. Controlling for these covariates is important while examining the influence of SHS exposure on birth weight.4.2Birth weight is inversely correlated with both neonatal and postnatal mortality. The increased risk is not limited to LBW infants. Infants with birth weights of 3,000–3,499 grams have infant mortality rates that are higher than those of infants weighing 3,500–3,999 grams at birth . However4.3.The goal of reducing LBW incidence by at least one third between 2000 and 2010 is one of the major goals in ‘A World Fit for Children’ . Physicaversus 9%) [Since nearly 1.1 billion people who smoked cigarettes consumed a total of five billion cigarettes in 1995, it is reasonable to assume that SHS exposure is a prevalent and an important public health problem. This is particularly true for developing countries, where men smoke substantially more than women (49% rsus 9%) . Swift arsus 9%) . This carsus 9%) .The workplace is also important to consider because women of childbearing age are present in the workplace in greater numbers than ever before. For example, women’s participation in the formal labor force in Jordan grew from 15% in 1980 to 24% in 1999 . MoreoveIn summary, SHS exposure is a prevalent and an important public health problem, especially in developing countries. Exposure to SHS plays an important role in determining the birth weight. Combining educational programs and legislative interventions aimed particularly at eliminating tobacco use is the best public health strategy to reduce SHS influence on birth weight. Moreover, application of antismoking regulations in the workplace is a key component of the required public health policies.4.4.Although the study sample has the limitation of only being able to be generalized to those women who give birth in the public hospital of Irbid, Jordan, it has the strength of being systematically selected from this population. In Jordan, 95% of the deliveries occur in hospitals. Although 65% of these deliveries occur in public hospitals, public hospitals do not represent the whole population, since 35 % of deliveries occur in private hospitals. Also the self-reported measures of exposure are likely to be imprecise and subject to recall bias.


Low birth weight is associated with an increased risk of neonatal and infant mortality and morbidity, as well as with other adverse conditions later in life. Since the birth weight-specific mortality of a second child depends on the birth weight of an older sibling, a failure to achieve the biologically intended size appears to increase the risk of adverse outcome even in babies who are not classified as small for gestation. In this study, we aimed at quantifying the risk of neonatal death as a function of a baby's failure to fulfil its biologic growth potential across the whole distribution of birth weight.We predicted the birth weight of 411,957 second babies born in Denmark (1979–2002), given the birth weight of the first, and examined how the ratio of achieved birth weight to predicted birth weight performed in predicting neonatal mortality.For any achieved birth weight category, the risk of neonatal death increased with decreasing birth weight ratio. However, the risk of neonatal death increased with decreasing birth weight, even among babies who achieved their predicted birth weight.While a low achieved birth weight was a stronger predictor of mortality, a failure to achieve the predicted birth weight was associated with increased mortality at virtually all birth weights. Use of family data may allow identification of children at risk of adverse health outcomes, especially among babies with apparently "normal" growth. Birth weight correlates with the risk of perinatal and infant mortality and morbidity -4, as weIn this study, we aimed at quantifying the risk of neonatal death as a function of a baby's failure to fulfil its predicted growth potential across the whole distribution of birth weight, including the "normal" range. To this end, we predicted the birth weight of 411,957 second babies born alive in Denmark between 1979 and 2002 using a modified version of the method proposed by Skjaerven et al .The Danish Civil Registration System (CRS) , which iDuring the study period, there were 1,476,753 live births to Danish residents, all of which are included in the Medical Birth Registry. We excluded 51,372 (3.5%) births due to missing information on birth weight or gestational age.Gestational age is often estimated with error -24, withUsing data on family members recorded in the CRS , we idenWe first calculated the predicted birth weight of second-born babies using the strategy described by Skjaerven et al , based oWe thus used the following equation to predict the birth weight of the second-born baby:wherePred(W2): predicted birth weight of the second-born child,I: estimated intercept among sibs where the first-born had a gestational age of g1 and the second-born had a gestational age of g2.β: estimated slope among sibs where the first-born had a gestational age of g1 and the second-born had a gestational age of g2w1: first-born baby's observed birth weightγ: estimated sex correction depending of the first-borns sex (s1) and the second-borns sex (s2)Note: All parameters were estimates simultaneously using a variance component model.g1) was categorized as 28–33, 34–35, 36–37, 38–39, 40–41, and >= 42 completed weeks, and second-borns' gestational age (g2) was categorized as 28–29, 30–31, 32–33, 34, 35, 36, 37, 38, 39, 40, 41, and >= 42 weeks. Since we expected gestational age of the second-born babies to be more important for predicting their own birth weight than gestational age of their older sibling, we decided a priori to use a more detailed categorization of gestational age for the second-born. For each gestational age of the first- and second-born, the parameters of intercept I and slope β are presented in Table γ is shown for each of the four combinations .In these analyses, first-borns' gestational age /(predicted birth weight) × 100]. The relative risk of death was estimated by log-linear Poisson regression treating the number of person-years as an offset variable [Second-born babies were followed from birth to the 27variable . All estth week of gestation, babies whose older sibling weighed 2000 grams had a mean birth weight of 3175 grams, while babies whose older sibling weighed 5000 grams had a mean birth weight of 4200 grams. These patterns were strikingly similar to those reported by Skjaerven et al [Figure en et al , who fouHowever, when we stratified the results shown in Figure To compare our prediction with that proposed by Skjaerven et al , we applAmong the 411,957 second-born children included in the study, 946 died within 27 days after birth, and 30 were lost to follow-up due to emigration from Denmark and were thus censored at the time of emigration.Within each category of the achieved birth weight, the risk of neonatal death depended on the birth weight ratio Table . In addiFigure Restricting the analysis to second-born babies born at term 37–41 weeks) whose older siblings were alive at their first birthday yielded similar results .In this study, failure to fulfil the estimated growth potential increased the risk of neonatal death. Several authors ,11,14,15The heterogeneity in risk of neonatal death observed among babies within the same weight category provides further evidence that birth weight alone is an incomplete marker of fetal growth although, used alone, the birth weight ratio would be an even poorer predictor at the lower birth weights. However, in the absence of clinical information, the birth weight ratio may be particularly useful in the "normal" range of birth weight.Unlike Skjaerven et al , we chosThe main limitation of this approach is that it requires an older sibling to make the prediction. In the absence of a sibling, other models have been proposed to predict birth weight, such as measures based on maternal characteristics ,30 or usWhen the birth weight of the first baby is extremely small or extremely high, due to either measurement error or natural variability, this will lead to attenuation of the relation between the birth weights of the two siblings. While this may reduce the ability of the model to predict birth weight, gestational age and birth weight of the older child still explained a large fraction (48%) of the total variation of the birth weight of the younger child.st birthday.When we examined the risk of neonatal death of the second baby as a function of how close the achieved birth weight was to the predicted birth weight (birth weight ratio), we observed an inverse J-shaped curve for all strata of achieved birth weight above 2000 grams; overall, mortality was higher for infants with a low birth weight ratio, lower for infants who achieved their predicted birth weight, and slightly higher for infants who achieved a birth weight higher than predicted. However, a low achieved birth weight was more strongly associated with a high risk of mortality, even among babies who fulfilled their prediction, and after adjustment for gestational age, which illustrates the limitations in making predictions in the absence of clinical data. Our results indicate that babies within the same category of birth weight have different risks of mortality, as previously suggested , dependiOur findings showed that a previous child allows a more individualized prediction of birth weight than gestational age alone. The second baby may differ genetically from the older sibling, but it has been suggested that maternal characteristics correlate with birth weight more closely than fetal genes . WhetherOur study spans a long time period, during which important improvements in the care of premature newborns have occurred. The relatively small absolute number of deaths in our study did not permit analyses of separate time periods. However, adjusting for year of birth did not change our estimates. Our results only refer to neonatal mortality in singleton live births. Results for post-neonatal mortality may be different. Our study is further limited by the lack of information on congenital anomalies or infections, as these conditions increase the risk of both growth restriction and neonatal death. This will limit the predictive value of our estimates, especially at the lower weights. As these conditions may be present in both pregnancies, this mechanism may, in part, explain the strong predictive risk of a small achieved birth weight among babies who appeared to have fulfilled their growth potential. Similarly, we did not have information on maternal morbidity in either pregnancy, and the mother's health status is likely to play an important role in growth.The main utility of the proposed approach lies, however, in its ability to provide a framework for assessing risk among babies with apparently "normal" birth weight. If individuals whose fetal growth was compromised have an increased risk for adverse health conditions, early identification may improve their outcome through monitoring or intervention. In the presence of detailed clinical information, the proposed method is unlikely to help physicians assess individual risk. However, a discrepancy in birth weight between siblings may constitute a warning sign even in the absence of obvious pathology, and such babies may benefit from increased surveillance. Babies who failed to achieve their predicted birth weight but were classified as "normal" by the regular criterion of small-for-gestation appeared to be at higher risk of delayed motor development . It is tOur results lend further credibility to the notion that impaired fetal growth is a marker of compromised development. While a low achieved birth weight was a stronger predictor of mortality, a failure to achieve the predicted birth weight was associated with increased mortality at virtually all birth weights. The approach described here may aid clinicians in the identification of babies at higher risk among those with a birth weight in the normal range. It should also be considered as a tool in epidemiologic studies that aim at studying medium- and long-term consequences of fetal growth disruptions.The author(s) declare that they have no competing interests.All authors contributed to the idea and conception of the study. CBP and OB suggested the initial design of the study. All authors contributed to the final design used. CBP analysed the data. All authors contributed to the interpretation of data. CBP and OB drafted the manuscript. All authors revised the manuscript critically for important intellectual content. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


We investigated childhood brain tumours by histological subtype in relation to prenatal X-ray among all children, less than 15 years of age, born in Sweden between 1975 and 1984. For each case, one control was randomly selected from the Medical Birth Register, and exposure data on prenatal X-ray were extracted blindly from antenatal medical records. Additional information on maternal reproductive history was obtained from the Medical Birth Register. We found no overall increased risk for childhood brain tumour after prenatal abdominal X-ray exposure (adjusted odds ratio (OR): 1.02, 95% confidence interval (CI): 0.64–1.62); primitive neuroectodermal tumours had the highest risk estimate . From 1975 to 2002, the incidence rate of childhood brain tumours (CBTs) in the United States increased from 2.3 to 3.5 cases per 100 000 children . A step Few risk factors are considered as established: rare genetic disorders and high-dose ionising radiation for cancer therapy and tinea capitis . RadioloDuring recent years, more focus has been on risk factors for the subtypes of CBT, since their cellular origins as well as geographic distribution differ. The few previously existing studies on prenatal X-ray and subtypes of CBT are based on retrospectively collected information and exposure frequency was low  to facilitate comparison over time. Cases were children less than 15 years of age, diagnosed with brain tumour . The histological subtypes are defined through the three-digit pathohistological diagnosis code assigned by a pathologist. Controls were randomly selected children from the Medical Birth Register and frequency matched to cases by gender and year of birth. Controls had to be alive and residents in Sweden until the age of 15 years, which was ensured by using information from the Cause of Death Register and the Register of Population and Population Changes.Medical information was abstracted from antenatal records, kept at local delivery archives and completed with data from the Medical Birth Register. Individual record linkage between registers and antenatal records was made possible by the Personal Identification Number (PIN), which is assigned to each Swedish resident and included in registers and medical records.The Medical Birth Register contains prospectively collected information on more than 99% of all births in Sweden since 1973  of at least 1.7. In the Cancer Register, we identified 601 children born between 1975 and 1984 with a diagnosis of brain tumour before the age of 15 years. For 62 of the 601 cases, the PINs were incomplete or information on hospital of birth was missing, which made it impossible to identify their antenatal records. Of the remaining 539 cases, we were able to retrieve antenatal records for 512 (95%). For the 539 control subjects, we found 524 records (97%).Exposure data were extracted blindly with regard to case or control status from both antenatal records and registers. All data were assessed by one of the authors (KS) before transfer to the database. From the antenatal records, we included information on pregnancy complications and other current diseases and data on X-ray exposure. From the Medical Birth Register, we retrieved information on maternal age and reproductive history, gestational age at birth, birth weight and mode of delivery. Information concerning smoking habits was collected from the Medical Birth Register (data included from 1982) and completed by data from the antenatal records.Radiological examinations were divided into abdominal and non-abdominal examinations. Abdominal examinations included pelvimetry, X-ray for foetal position and others. Risk for CBT was evaluated for all types of brain tumours combined and according to subtype. Each diagnosis was classified into one of the following subtypes: low-grade astrocytoma , high-grade astrocytoma (astrocytoma grades III–IV), primitive neuroectodermal tumour (PNET) defined as category IC–D in the modified WHO classification by We included information on numbers of X-ray examinations and the gestational week when they were performed. Gestational age was divided into trimesters: the first , second (weeks 15–28), and third (weeks 29–45). Gestational age was calculated from data on last menstrual period, which was the standard procedure in Sweden at this time . Birth wLogistic regression was performed to evaluate the association between prenatal exposure to diagnostic X-ray and incidence of CBT. Estimates of ORs and 95% confidence intervals (95% CIs) were calculated. In advance, we identified potential confounding factors that could interfere with both exposure and outcome, and for which it was possible to obtain information either from the registers or the antenatal records. The following confounders were included in the adjusted analyses: maternal age at birth, parity, multiple birth, mother's country of birth  or non-Nordic country), hypertension during pregnancy, mode of delivery, breech position, gestational age at birth, birth weight, head circumference at birth and level of hospital .The statistical analyses were conducted with the SAS 9.1 software package.The study was approved by the Ethics Committees at Karolinska Institutet and Uppsala University.vs 37% smokers, respectively). However, the information on smoking habits was missing for 36.5% of the sample. Further maternal and foetal characteristics are presented in P=0.01) and to be born at a primary or secondary level hospital (P=0.04) than for controls. There were no other significant differences in maternal and neonatal characteristics between cases and controls  of the mothers to children diagnosed with CBT were exposed to X-ray during pregnancy compared to 21.2% (n=111) of mothers to the children randomly selected as controls. For abdominal X-ray, the exposure frequency was 10.7% (n=55) for cases and 9.4% (n=48) for controls. Beside pelvimetry and X-ray for foetal position, abdominal X-ray included three scintigraphy of placenta, four scintigraphy of kidney, four plain abdominal X-rays and one colon examination. For both cases and controls, more than 96% of the abdominal examinations were performed in the last trimester. Non-abdominal examinations included 107 pulmonary and one each of skeletal, sinus and dental X-ray. Only three individuals in each group were exposed to X-ray more than once during the pregnancy.Overall, 21.1% (n=17) were included in all cancers and not analysed separately.In Prenatal X-ray exposure does not seem to increase the overall risk of CBT. However, for the subgroup of PNET, we found an increased OR of 1.88 (95% CI: 0.92–3.83). PNETs are found among the youngest children and is classified as an embryonic tumour. Medulloblastoma is the most common tumour in the PNET group, has a peak age of occurrence of 7 years of age and is located in the cerebellum. The other, and more rare tumours in the PNET group, may occur in the neonatal period , have a peak age at diagnosis of about 5 years of age and are mainly supratentorial , and antenatal pelvimetry is rarely indicated. Nevertheless, abdominal radiation still occurs, for instance in CT scan on maternal indications. Risks with foetal diagnostic radiation could probably be extrapolated to the early childhood period, since the brain still develops and grows rapidly during the first years of life.Our most important finding might not be the clinical implication, but in relation to the theory that prenatal or neonatal radiation might affect the various neural cell types differently. PNET are presumed to arise from undifferentiated neural stem cells and little is known regarding the molecular genetic events initiating carcinogenesis . An inteIn conclusion, we found no increased risk for CBT after prenatal X-ray exposure when analysed as one group. When analysing according to subgroups, the embryonic tumours defined as PNET showed the highest risk estimates.


The aim of the study was to report the incidence of retinopathy of prematurity (ROP) and severe ROP and identify the risk factors for their development in a large nursery in Kuwait.This was a retrospective, interventional, non-comparative, hospital-based study. Retrospective review of ROP records of premature babies having either birth weight of less than 1501 g or gestational age at birth of 34 weeks or less and born between January 2001 and August 2003.By univariate and multivariate logistic regression analysis.P < 0.001), gestational age , surfactant  and stay in the intensive care unit for longer than 15 days  to be significant for the development of any ROP. Low birth weight , bacterial sepsis  and need for surfactant  were found to be the risk factors for severe ROP needing laser treatment.Out of the 599 babies studied, 38.9% developed ROP and 7.8% needed treatment for severe ROP. Multivariate analysis showed low birth weight (OR 13.753, 95% CI 3.66-51.54; (The incidence of both any ROP and ROP needing treatment are comparable to other studies. Low birth weight is the most important risk factor for both any ROP and severe ROP. Retinopathy of prematurity (ROP) is a well-established disease of premature babies and has emerged as an important cause of childhood blindness.2 After tThe risk factors for the development of ROP and threshold ROP include low birth weight,710–12 lo7565610677There are only a few studies from the Arabian Gulf region regarding the incidence of ROP and the risk factors for the development of ROP.15–17 TheIn a retrospective, interventional, non-comparative, single-center case series study, we reviewed the medical records and ROP charts of premature babies who were born between January 1, 2001 and August 31, 2003 at the neonatal intensive care unit, Al Sabah maternity hospital, Kuwait city, Kuwait. Approval was obtained from the Institutional Ethics Committee for the study. Efforts were made to remain true to the guidelines of the Declaration of Helsinki Principles.Babies either with a birth weight of less than 1501 g or born at a gestational age of 34 weeks or less were screened for ROP by ophthalmologists if they survived up to 28 days. The first examination was conducted at four weeks of chronological age. Follow-up examination was done every two weeks in the absence of active ROP and weekly or earlier if active ROP was detected. In this study we treated babies when the ROP reached Stage 3 in Zone I or II with 3 clock h of contiguous or 5 clock h of cumulative extraretinal proliferation in the presence of plus disease. We termed this stage as modified threshold disease (MTD) in our study. Babies who did not develop ROP or those who developed acute ROP but did not reach MTD were followed up till the retina became fully vascularized or achieved 50 weeks of gestational age. Dilatation of the pupils was achieved with cyclopentolate 0.5% eye drops instilled three times at 15 min-intervals prior to the screening examination. The findings of indirect ophthalmoscopy were recorded in a ROP datasheet at each visit.Acute ROP was classified according to the international classification of retinopathy of prematurity (ICROP) study recommendations. Babies wData were collected for each baby regarding date of birth, sex, single or multiple pregnancy, gestational age at birth, birth weight, surfactant given or not and presence of common problems of prematurity like intraventricular hemorrhage (IVH), bacterial or fungal or combined infection, hyaline membrane disease (HMD) and its severity, necrotizing enterocolitis (NEC) and patent ductus arteriosus (PDA) and its treatment. Presence of intrauterine growth retardation (IUGR) and duration of stay of the infant in the intensive care unit of the neonatology (NICU) department were also noted.P value of 0.05 or less was considered statistically significant. Odds ratio and 95% confidence intervals were calculated for determining the independent risk factors.All data was entered into a spreadsheet and analyzed using SPSS for Windows statistical software Version 10.0 . Univariate and multivariate logistic regression was done to determine the risk factors for the development of any ROP and MTD. A There were 624 babies eligible for screening for ROP. However, 13 babies died before the completion of screening and another 12 babies did not follow up fully for screening examination. After excluding these 25 babies, we identified 599 babies who had completed the examination.There were 313 (52.3%) male and 286 (47.8%) female babies. The average birth weight of 599 babies was 1270.01 g . The average gestational age was 31.1 weeks . Three hundred and seventy-four babies (62.4%) were the product of single pregnancy and 225 babies (37.6%) were product of multiple pregnancies. The 225 babies included 116 twins (19.7%), 66 triplets (11.1%), 38 quadruplets (6.3%) and five quintuplets (0.8%).A significant number of babies had a stormy neonatal course with IVH in 84 (14.0%), proven bacterial sepsis in 121 (20.2%), fungal infection in 27 (4.5%) and mixed fungal and bacterial infection in 12 (2.0%) babies. Nearly 50% of babies (295) had HMD of varying degree. Administration of surfactant was needed in 353 (58.9%). PDA needed administration of indomethacin treatment in 43 babies (7.7%) and surgical ligation in 10 babies (1.67%). IUGR was present in 42 (7.0%) and NEC in 20 babies (3.3%).Some ROP was detected in 233 babies (38.9%) in at least one eye and the remaining 366 61.1%) babies did not develop any ROP. Bilateral ROP was detected in 208 (34.7%) and unilateral disease was found in 25 babies (4.2%). The highest stage of ROP reached, in at least one eye in these babies, is shown in Tables .1% babieThe average birth weight of babies developing ROP and MTD was 1077.73 g  and 874.68 g  respectively. The average GA of babies developing ROP and MTD was 29.66  and 27.74  weeks respectively.Eighty-six eyes of 47 (7.8%) babies were found to have MTD or worse and were treated by laser ablation. Bilateral disease was seen in 39 (6.5%) babies and unilateral disease was seen in eight (1.3%) babies. Seventy-nine eyes of 43 babies had MTD in Zone II and seven eyes of four babies had MTD in Zone I. Thirteen eyes needed retreatment for skipped areas. All the unilateral cases showed good regression of MTD after laser treatment, but five of 39 cases with bilateral disease showed progression of disease even after laser treatment. One baby, who showed bilateral progression to Stage 4A had spontaneous regression in both eyes. In another baby, Stage 4A in one eye was successfully treated by scleral buckling and the other eye had spontaneous regression of Stage 4A. Two babies showed progression to Stage 5 in both eyes and parents refused surgery. The fifth child progressed to Stage 4B in both eyes and underwent pars plana vitrectomy in both eyes. The child had successful re-attachment of the retina in one eye. In summary, three children (6.4% of treated babies) had unfavorable outcome in at least one eye. Unfavorable structural outcome was defined as total retinal detachment or presence of macular fold.[P < 0.001), GA (P < 0.001), presence of IVH (P < 0.001), need for surfactant (P < 0.001), presence of bacterial infection (P < 0.001), presence of fungal infection (P < 0.001), need for stay in NICU for more than 15 days (P < 0.001), presence of HMD (P < 0.001) and presence of NEC (P < 0.001) to be significant for the development of any ROP. Univariate analysis showed birth weight (P < 0.001), GA (P < 0.001), presence of IVH (P < 0.001), need for surfactant (P < 0.001), presence of bacterial infection (P < 0.001), presence of fungal infection (P < 0.001), presence of combined fungal and bacterial infection (P < 0.001), need for stay in NICU for more than 15 days (P < 0.001), presence of HMD (P < 0.001), presence of PDA(P = 0.005) and presence of NEC (P < 0.001) to be significant for the development of MTD. However, many of these factors were not found to be significant when subjected to more rigorous multivariate analysis. The risk factors identified by the multivariate analysis are summarized in Univariate analysis showed birth weight . Tables Low birth weight,710–12 lo757Multivariate analysis found low birth weight, bacterial sepsis and surfactant to be significant for the development of MTD . Low bir612et al.[et al.[et al.[The outcome of treated babies was satisfactory in our study with unfavorable outcome seen in 6.38% of the treated babies. The CRYO- ROP studies et al. who noteet al. showed tl.[et al. in a ranl.[et al. were publ.[et al. depends l.[et al. showed tThe incidence of both any ROP and ROP needing treatment in our study was comparable to other studies. Birth weight is the most important risk factor for both any ROP and severe ROP. Treating babies at the stage of MTD has yielded low incidence of unfavorable outcome in our study.


The aim was to assess the impact of nutritional status and environmental exposures on infant thymic development in the rural Matlab region of Bangladesh.max 2094 infants born during a randomized study of combined interventions to improve maternal and infant health, thymic volume  was assessed by ultrasonography at birth and at 8, 24 and 52 weeks of age. Data on birth weight, infant anthropometry and feeding status were also collected.In a cohort of NAt all ages, TI was positively associated with infant weight and strongly associated with the month of measurement. Longer duration of exclusive breastfeeding resulted in a larger TI at 52 weeks. TI at birth and at 8 weeks correlated positively with birth weight, but by 24 and 52 weeks and when adjusted for infant weight this effect was no longer present. Thymic size was not affected by pre-natal maternal supplementation or by socioeconomic status but was correlated to arsenic exposure during pregnancy.In this population of rural Bangladeshi infants, thymic development is influenced by both nutritional and environmental exposures early in life. The long-term functional implications of these findings warrant further investigation. Early infancy is a time of rapid immunological development, where antigenic stimuli can shape the developing immune system, potentially ‘programming’ long-lasting immune responsiveness. In a developing country context where the primary focus of research and health interventions is on infectious disease control, understanding this early-life programming is critical. Recent research suggests that nutritional status in early life may impact on human immune development with, for example, a positive association observed between birth weight and antibody response to certain vaccines later in life –3. The pThe human thymus is a primary lymphoid organ essential for the establishment of a normal peripheral T-lymphocyte immune system. In the thymus, bone marrow-derived precursors undergo differentiation and selection, ultimately leading to the migration of positively selected thymocytes to the T-cell-dependent areas of the peripheral lymphoid organs ,5. It haPost-natal development of the thymus can be assessed sonographically using a validated method to estimate a volume-related thymic index. This technique has previously been used to indicate that the thymus is sensitive to early mode of feeding, with a smaller mean thymic index observed in bottle-fed than breast-fed infants ,8. FurthIn rural Bangladesh, the MINIMat study has explored combinations of pre- and post-natal nutritional interventions to address the issues of maternal, fetal and infant malnutrition. In a sub-cohort of infants from the main study, we have used serial ultrasonography to explore early-life predictors of thymic development. As the concentration of arsenic in drinking water was frequently elevated in the study area, we also investigated the effects of arsenic exposure on thymic development.The study was conducted in the Matlab region of Bangladesh, a rural but densely populated riverine area, situated about 60 km southeast of the capital Dhaka. The population of the Matlab region is well characterized, having participated in a demographic surveillance system (DSS) since 1963 as part of a health-related research programme led by the International Centre for Diarrhoeal Disease Research, Bangladesh . In 2001, ICDDR,B initiated the ‘Maternal and Infant Nutrition Interventions, Matlab’ (MINIMat) study, randomizing all pregnant women in Matlab to receive a combination of protein energy and micronutrient supplements. On enrolment, usually around 9 weeks of gestation, women were randomly allocated to a prenatal food supplement in combination with three separate micronutrient supplements. At the time of the study in Bangladesh, an ongoing government-supported national programme issued a protein-energy dense supplement providing 600 kcal/day to pregnant women 6 days per week. On enrolment to the MINIMat study, women were randomized to start this food supplement early in the first trimester of pregnancy (early start) or in the second trimester as the usual national programme suggests . Women were then further randomized to three micronutrient supplements: (i) the UNICEF/UNU/WHO preparation of 15 different micronutrients including 30 mg iron and 400 μg folic acid, (ii) 60 mg iron and 400 μg folic acid or (iii) 30 mg iron and 400 μg folic acid daily from week 14 during pregnancy. All supplements continued up to birth. Peri-natally, women were further randomized to receive exclusive breastfeeding counselling or the standard health counselling provided by local caregivers. The prenatal arm of the study was completed in June 2004 with a total of 3267 singleton infants born. The full protocol and main findings from the MINIMat study will be published elsewhere. For the purpose of the current study, we recruited mothers and their infants from the main MINIMat trial either at birth  or at follow-up .For health centre deliveries, birth weight, length, head circumference and knee–heel lengths were measured immediately following delivery. For births occurring at home, a birth notification system was established to ensure that study staff were made aware of births as soon as they occurred. A female paramedic then visited the newborn within 72 h of birth. Birth weight was measured by electronic or beam scales, to the nearest 10 g. Recumbent length was measured using locally manufactured, collapsible length boards, precise to the nearest 1 mm.Following delivery, infant weight and length were measured at monthly visits till 12 months of age using regularly validated standard equipment. At separate monthly visits, a female fieldworker visited each household to collect information on infant feeding practices. Full details are as described elsewhere . Data onAs the study area had a high prevalence of tube wells with elevated concentrations of inorganic arsenic (As) , we alsoThymus size was assessed sonographically using a validated method in which the transverse diameter of the thymus and the sagittal area of its largest lobe are multiplied to give a volume-related thymic index (TI) . This inTI was assessed at four time points during infancy: within 24 h of birth  and at 8, 24 and 52 weeks of age. Infants were scanned in the supine position, using a transternal approach; sonography was not performed while the infant was crying. Owing to a delay with the start of TI assessment, a large proportion of infants had passed 16 weeks by the time the study started; this resulted in more infants with TI measured at the latter two time points.The main MINIMat study and the thymic size addendum were approved by the Research Review and Ethical Review Committees, ICDDR,B, Dhaka, Bangladesh. The thymic size addendum was additionally approved by the Ethics Committee at the London School of Hygiene and Tropical Medicine. Written informed consent was obtained from all participating mothers following an explanation of the study in their own language.t-tests. Associations between TI at each time point were assessed by correlation analysis, with TI measurements adjusted for season, weight and sex of the infant. TI was regressed on these covariates, separately at each time point, and the correlations between time points were calculated for the resulting residual . The effect of seasonality on TI size was assessed by fitting a truncated Fourier series regressed against TI (i) and cos(θi)] was considered. The effect of birth size on TI at each time point was assessed by multiple regression analysis, controlling for sex, infant length and season of measurement. Infant length was included in the models since thymic size is expected to be related to infant size. Thus at birth, the relationship between TI and birth weight should be interpreted in terms of weight-for-length. Since infants’ lengths tend to track along their starting centile, current size will be strongly correlated with birth length, and the TI–birth weight relationship will also have a similar interpretation at subsequent time points. For seasonality, the first two pairs of Fourier terms were fitted, i.e. the terms in β3 to β6, in the following model (2} to the model shown above). SES fitted as a five-level categorical variable was added to the model in order to check for confounding with [As].Comparisons among group means were made using two-sample ainst TI . To testng model :(1) In oF-test and the likelihood ratio test for ordinary multiple regression and multi-level models, respectively.In the regression analyses, models were compared using the ANOVA Z-scores were calculated using the Child Growth Standards of the WHO, using the WHO Anthro 2005 software and macros (http://www.who.int/childgrowth/software/en/). All statistical analyses were performed using DataDesk, version 6 for Windows from Data Description Inc.  or Stata 9 from Stata Corp. .2 (18.3–31.2), respectively, and a strong positive association was observed between both measures and infant weight at birth (p ≤ 0.0001 for both). A total of 1168 infants had their TI measured within 24 h of birth. When compared with infants who did not have TI measured at birth, infants who had TI measured at birth were significantly heavier  and had a significantly longer gestational age . No differences were observed with birth length .Anthropometric characteristics of the infants at each of the time points when TI was assessed are detailed in Mean (95% CI) TI at birth was 27.6 (27.1–28.0) and was positively correlated to birth weight and birth length . A significant but weak positive correlation was observed between TI at birth and both maternal weight and BMI at 30-week gestation  but not for parity (p = 0.5161). However, all other associations became non-significant once adjusted for infant birth weight.As a consequence of the delay in starting these measurements, the number of infants with TI assessed at each time point during infancy increased from 1715 and 1704 at 8 and 24 weeks, respectively, to 2094 infants at 52 weeks. Mean TI increased rapidly from birth to week 8, reaching maximum size at week 24 and then declining to the week 52 measurement . TI was 2 on 6 df = 21.18; p = 0.0017). This difference was almost entirely due to the difference between the pattern at birth and the patterns observed post-natally  while no further difference was detected between the post-natal patterns . The seasonal pattern with the time points pooled is also illustrated in The effect of month of measurement on mean TI at each time point and for all time points pooled is illustrated in Using an intention-to-treat analysis, we also explored the relationship between pre-natal supplementation and exclusive breastfeeding counselling and post-natal thymic development. No associations were observed between any of the prenatal interventions (food or micronutrients) and TI at any time point (data not presented). According to lactation counselling group, no associations were observed between TI at birth and at 24 or 52 weeks, but at 8 weeks, infants born to mothers who had received counselling in exclusive breastfeeding had a marginally significantly smaller TI than those who had not . Using mother-reported duration of exclusive breastfeeding (assessed in blocks of 15 days) as the exposure variable, no associations were observed with the measures of TI taken at 8 and 24 weeks. However, at 52 weeks, infants who had been exclusively breastfed for greater than the median duration of days of exclusive breastfeeding (59.9 days) had a significantly larger thymus than those who had not . This remained the case when adjusted for infant length and sex (p = 0.0129).2 = 12.93 on 2 df; p = 0.0016): TI was negatively associated with As, but this association diminished at high-exposure levels. Overall, however, As exposure explained only 4% of the variance in TI. Since the diminishing effect of As was unexpected, we considered the possibility that it might have resulted from confounding with SES. SES did correlate strongly with both TI and As, although the association with TI was entirely explained by its association with infant size. When added to the regression model, already controlled for infant length, SES had a negligible effect on the relationship between As and TI.The median As concentration in maternal urine for the 1556 mother–infant pairs who had both As data and TI measured at week 52 was 102 μg/L (range 5.5–1150 μg/L). The analysis of the effect of prenatal As exposure on TI revealed a complex relationship. The simple linear trend alone was not significant (p = 0.74) but the addition of the quadratic term gave a highly significant fit -7 levels , suggestThe region of Matlab represents an area of Bangladesh with a high prevalence of arsenic-contaminated tube well water . ChronicThe latter would be compatible with a prenatal effect on immune function and we are currently investigating child mortality in relation to prenatal arsenic exposure in more depth. Although we have recently shown that breastfeeding protects against early-life exposure to arsenic , non-excin utero influence on thymic development, possibly mediated by maternal nutritional status. Recent research has focused on the hypothesis that nutritional status during fetal life and early infancy may be critical for immune development, with long-lasting effects on functional immunity . We did however observe a positive correlation between birth weight and thymic size, an effect that persisted until the infants were 8 weeks of age, irrespective of the current infant size. This finding suggests an immunity . In Filiimmunity  and Pakiimmunity ,3, for eA limitation of the current study is the reliance on thymic size as an indicator for early immune development with no direct assessment of thymic function included. Little published data exist to correlate thymic size to function during infancy. In a small study of Danish infants, a positive correlation was observed between thymus size and the proportion of CD8+ cells in peripheral blood and the CD4/CD8 ratio at 10 months of age . In a suIn conclusion, we have shown that the human thymus is sensitive to a number of nutritional and environmental exposures and that these effects appear to persist throughout early infancy. The long-term consequence of having a smaller thymus are yet to be fully understood, although in a study from Guinea-Bissau, West Africa, a smaller TI at birth was associated with increased infant mortality  suggesti


Early-life exposures to hormonally active compounds and other factors may affect later response to estrogen or progesterone and hence may influence development of uterine leiomyomata (fibroids).in utero and early-life exposures, including soy formula, with self-report of physician-diagnosed fibroids by 35 years of age.We evaluated associations of Our study included 19,972 non-Hispanic white women who were 35–59 years of age when they enrolled in the Sister Study in 2003–2007. We estimated risk ratios (RRs) and 95% confidence intervals (CIs) using log-binomial regression models for fibroid associations with adjustment for participant’s age and education, maternal age at participant’s birth, birth order, and childhood family income.In utero diethylstilbestrol (DES) exposure was also associated with early fibroid diagnosis , but this association was driven by women reporting probable rather than definite exposure.Greater risk of early fibroid diagnosis was associated with soy formula during infancy , maternal prepregnancy diabetes , low childhood socioeconomic status , and gestational age at birth . in utero DES, in relation to fibroid risk, and replication of findings in other populations is needed.There are plausible biological pathways by which these early-life factors could promote fibroid pathogenesis. This is the first epidemiologic study to evaluate such exposures, with the exception of  Uterine leiomyomata (fibroids) are benign smooth-muscle tumors, which are associated with pelvic pain, heavy bleeding, and reproductive problems , and thiin utero diethylstilbestrol (DES)  , age at ol (DES) , and chiol (DES) .in utero, early-life, and childhood exposures were associated with self-report of fibroid diagnosis by 35 years of age among non-Hispanic white participants in the NIEHS Sister Study. Specifically, we evaluated the following participant factors: birth weight, gestational age at birth, birth order, being from a singleton or multiple birth, breast milk or soy formula consumption during infancy, relative height and weight during childhood, age at menarche; and the following childhood socioeconomic factors: maternal education, highest level of education in the household, food insecurity, and relative family income. In addition, we evaluated maternal factors related to the pregnancy with the participant: age at birth, living and working on a farm, smoking during pregnancy, prepregnancy diabetes, DES use, and complications of pregnancy .We examined whether The NIEHS Sister Study  is a proParticipants were mailed kits containing three questionnaires for self-completion , written consent documents, and support information for telephone interviews and home visits. A prepaid phone card was included to encourage participants to contact their mother or other relatives about early-life events and family medical history. Home visits included measurement of participants’ height and weight and retrieval of completed questionnaires. Participants completed a computer-assisted telephone interview to collect information on known and suspected risk factors for breast cancer and other end points. The Sister Study was approved by the Institutional Review Board of the NIEHS, National Institutes of Health, and all participants provided their informed consent.Baseline data were available for 32,071 women who completed their interview and home visit by 21 September 2007. We restricted our analysis to 20,061 non-Hispanic whites who were 35–59 years of age when fully enrolled into the Sister Study. We excluded women > 59 years of age because of the likelihood of secular differences in use of ultrasounds for fibroid diagnoses and decreased ability to get information from their mothers about early-life events. Because of racial/ethnic differences in fibroid risk and related morbidity , we restDuring the baseline interview, women reported whether they were ever told by a doctor or health professional that they had uterine fibroids and their age at time of diagnosis. Because ultrasound screening has shown that the prevalence of fibroids increases rapidly after 35 years of age in U.S. white women , many olWe used self-administered family history questionnaires to assess most intrauterine and early-life exposures: birth weight (pounds/ounces), gestational age at birth (categories), singleton or multiple birth, breast milk or soy formula consumption during infancy; and maternal exposures: age at birth, living and working on a farm, smoking during pregnancy, prepregnancy diabetes, DES use, and complications of pregnancy . Except for questions on gestational age, birth weight, singleton or multiple birth, and maternal age at birth, the response categories were “definitely,” “probably,” “probably not,” and “definitely not” to allow for possible uncertainty about early-life events. Questionnaires also assessed the duration that participants were fed with breast milk or soy formula and whether they were fed with soy formula during the first 2 months of life. However, we did not evaluate duration of breast-feeding or soy formula in relation to fibroids because of a considerable proportion of missing responses among women who reported being breast-fed (> 20%) or fed soy formula (> 30%). For analyses, we did not consider mothers as having gestational diabetes if they were reported to have diabetes before pregnancy. If participants reported that their mothers had both preeclampsia and pregnancy-related hypertension, we considered their mothers as having preeclampsia in analyses, and we considered mothers as having pregnancy-related hypertension only if preeclampsia was not also reported.Birth order was estimated from birth dates of brothers reported on the family history questionnaires and sisters reported during the computer-assisted telephone interviews. We included all full siblings and half-siblings who shared the same mother for defining birth order. Childhood exposure data—maternal education and maximum level of education for parents or guardian in the household when participant was 13 years of age, relative family income based on self-reported categories , not enough to eat at times during childhood, age at menarche, and height and weight relative to peers at age 10—were collected during the telephone interview. Participant’s age, highest level of education, smoking status, alcohol intake, parity, and menopausal status were also assessed during the telephone interview, and body mass index was calculated using weight and height measured at the home visit.All statistical analyses were conducted using SAS . We used log-binomial regression to estimate risk ratios (RRs) with 95% confidence intervals (CIs) for associations between each of the intrauterine, early-life, and childhood exposures and fibroid diagnosis by 35 years of age. For each of the relevant exposures, we combined women who reported definite or probable exposure for estimation of RRs and considered women unexposed if they reported probably not or definitely not. We later performed sensitivity analyses in which we estimated associations separately for women who reported definite exposure and those who reported probable exposure. We estimated associations with fibroids within the following categories of birth weight: < 2,500 g, 2,500–2,999 g, 3,000–3,499 g, 3,500–3,999 g, and ≥ 4,000 g. For maternal age, we initially examined associations with 5-year age categories and then further combined categories that were associated with a similar risk of fibroids. Likewise, we initially explored fibroid associations across all categories of birth order and maternal and household education, and then combined categories with similar fibroid risk for reporting of associations. Although approximately 40% of women did not report gestational age at birth, we evaluated its association with fibroids, categorizing women as having been born at least 1 month early, 2–4 weeks early, and not early . Birth weight analyses were repeated with exclusion of women who reported being born at least 1 month early to further our understanding of birth weight and gestational age associations. We estimated RRs that were adjusted for participant factors (participant’s age and education) that may affect reporting of fibroids and exposures. We also adjusted RRs for early-life factors  that may be associated with other exposures and fibroids but were not considered to be on the causal pathway. The sample sizes for age-adjusted and fully adjusted models shown in tables differ. Although these differences are only about 2%, we repeated the age-adjusted analyses for the 19,531 women with complete data on variables included in the fully adjusted models and verified that results were similar to those for the complete sample of 19,972 women (data not shown).The prevalence of self-reported fibroids diagnosed at any age was 25% (data not shown), with 8% of women reporting early diagnosis by 35 years of age . Women rFive of the 14 early-life factors we examined were associated with more than a 20% increase in risk of early fibroid diagnosis  . We notein utero exposure to DES  and gestational diabetes  were associated with fibroids, but there were no associations with definite in utero exposure to DES and gestational diabetes  and fibroids was attenuated in the fully adjusted model. After repeating our analyses by estimating fibroid associations separately for women who reported “definitely” and those who reported “probably” for relevant early-life exposures, reporting of probable diabetes .Childhood socioeconomic factors, including less than high school for highest level of household education, not enough to eat at times during childhood, and being poor in childhood, were associated with fibroids, with the strongest association for being poor  . BecauseWe estimated an increased risk of early fibroid diagnosis in association with being fed soy formula during infancy, having a mother with prepregnancy diabetes, being born at least 1 month early, and reporting factors indicating low socioeconomic status during childhood . We also noted associations with fibroids for having a mother with gestational diabetes and DES use during pregnancy, although these associations were restricted to reporting probable rather than definite exposure.The association with soy formula is of interest given the estrogenic isoflavones found in soy products. Infants fed soy formula are exposed to isoflavone levels that are more than five times higher than typical levels for adults consuming soy-based foods . GenistiGenistein has been extensively investigated in laboratory animals. In particular, neonatal treatment of mice with genistein has been associated with development of uterine adenocarcinoma , abnormain utero exposure to maternal diabetes affects fibroid risk in early adulthood. One hypothetical mechanism by which in utero exposure to diabetes would affect later fibroid pathogenesis is the alteration of methylation patterns in regions that affect expression of relevant genes. In particular, expression of imprinted genes is especially sensitive to changes in methylation patterns . These factors may influence development of fibroids through changes in methylation patterns in childhood that persist and affect gene expression as adults. Plausibility of this hypothesis is based on animal and human data on early-life neglect or abuse. Different methylation patterns within the hippocampus of adult rats were detected based on whether they received maternal care early in life . In a smWe also found a strong association with fibroids for being born at least 1 month before mother’s due date. A weak association with low birth weight was no longer present after excluding women who were born at least 1 month early. Because levels of estrogen and progesterone rise throughout pregnancy, one hypothesis is that women who are born early are deprived of the estrogen needed for full differentiation of their reproductive system . Our anaSelection bias is a potential limitation for other exposures as well, given that the proportion of missing values was as high as 20%. There were slight differences in the proportion with fibroids based on whether exposures were missing, with generally more women reporting fibroids among those with missing responses. Based on the assumption that women with missing data for rare exposures  were likely to be unexposed, we repeated analyses in which we considered women with missing values for these factors as unexposed. However, changes in RR estimates were minimal (data not shown).in utero DES exposure and maternal gestational diabetes, for which associations were much stronger with reporting probable exposure. Because none of these exposures is known to be related to fibroids, exposure misclassification would likely be nondifferential, which would generally result in RR estimates biased toward the null.There is also the potential for misclassification of exposures given that women were reporting exposures during infancy and related to their mother’s pregnancy. However, women were provided phone cards to encourage them to ask these questions directly of their mothers, and we excluded older women (> 59 years) who would be less likely to have living mothers to ask about these exposures. In addition, response categories for many of the exposures included options of “definite” and “probable” that allowed for uncertainty in reporting. Associations with fibroids were generally consistent for definite and probable exposure, except associations with Our assessment of fibroid diagnoses was based exclusively on self-report. However, because fibroid incidence increases strongly with age, we only considered diagnoses by 35 years of age to reduce misclassification in the noncase group. Our estimated risk of 8% for early diagnosis of fibroids was similar to the risk of 11% for self-reported fibroids by 35 years of age in white women (35–49 years of age) from the NIEHS Uterine Fibroid Study . We also reported a substantially greater proportion of women with early diagnosis of fibroids having hysterectomies compared with women without early diagnosis, which suggests that many of the women with fibroid diagnoses included in our case definition had fibroid-related morbidity. However, we do not have information on fibroid-related symptoms at time of diagnosis. We also excluded older women from our analyses because of possible secular differences in use of ultrasounds for fibroid diagnoses. Because studied factors may also be related to fibroids diagnosed later in life, including women with fibroids diagnosed after 35 years of age as noncases may have resulted in an underestimation of RRs. However, repeating our analyses after exclusion of women with later diagnoses of fibroids (> 35 years) from the noncase group did not affect RR estimates with early diagnosis of fibroids for four of our main findings .Strengths of this study include a large sample size, which allowed us to examine associations with rare intrauterine and early-life exposures. We adjusted for factors that may affect recall of exposures, including participant’s age and education. In addition, despite the potential for misclassification bias from self-reported exposure information and fibroid diagnoses, we observed expected associations between specific factors, including a positive association between early age at menarche and fibroids and an increased reporting of maternal preeclampsia among firstborn women and those from a multiple birth (data not shown).Our study suggests that being fed with soy formula during infancy, having a mother with prepregnancy diabetes, being born at least 1 month early, and growing up with low socioeconomic conditions may increase the development of fibroids in early adulthood. This is the first study to explore these early-life and childhood factors in relation to the risk of fibroids. There are plausible biological mechanisms by which these factors could affect uterine physiology later in life and thus increase risk of fibroid development. Replication of findings in other populations including higher risk groups such as African Americans is needed.In


The extracellular matrix (ECM) as well as soluble mediators like cytokines can influence thebehavior of cells in very distinct as well as cooperative ways. One group of ECM moleculeswhich shows an especially broad cooperativety with cytokines and growth factors are the proteoglycans.Proteoglycans can interact with their core proteins as well as their glycosaminoglycanchains with cytokines. These interactions can modify the binding ofcytokines to their cell surface receptors or they can lead to the storage of the soluble factors inthe matrix. Proteoglycans themselves may even have cytokine activity. In this review wedescribe different proteoglycans and their interactions and relationships with cytokines andwe discuss in more detail the extracellular regulation of the activity of transforming growthfactor-β (TGF-β) by proteoglycans and other ECM molecules. In the third part the interactionof heparan sulfate chains with fibroblast growth factor-2  as a prototypeexample for the interaction of heparin-binding cytokines with heparan sulfate proteoglycansis presented to illustrate the different levels of mutual dependence of the cytokine networkand the ECM.


Cystic ovarian teratomas comprise 20% of all ovarian neoplasms, and are commonly encountered in patients between 20 and 40 years of age. Although these cysts are usually asymptomatic, we present the case of a patient whose cyst resulted in pruritus and abdominal pain. Based on a MEDLINE search of the literature, we believe this is the first case report of a twisted ovarian cyst presenting with generalized pruritus.A 35-year-old Sri Lankan woman presented with lower abdominal pain of one day’s duration with vomiting and generalized pruritus. She had no history of allergies and was not on medication. Upon a physical examination, our patient was found to have an acute abdomen, localized peritonism in her lower abdomen and tachycardia of 100 beats per minute. Computed tomography showed that the cyst, which contained calcified structures, originated from her left ovary. After laparoscopy-assisted removal of the twisted ovarian cyst, her symptoms resolved completely. Histological examination confirmed a benign ovarian teratoma.An unusual case of torsion of an ovarian teratoma presenting with abdominal pain and generalized pruritus, believed to be due to an antibody-mediated response, was resolved after surgical removal of the cyst. Mature cystic ovarian teratomas comprise 20% of all ovarian neoplasms and are encountered in the second or third decade of life . These cA 35-year-old Sri Lankan woman, previously well, presented with lower abdominal pain of one day’s duration with vomiting and generalized pruritus. She did not report a history of previous allergies and was not on medication at the time of presentation. An examination revealed an acute abdomen, localized peritonism in her lower abdomen and tachycardia of 100 beats per minute. Hematological investigations revealed a normal full blood count and normal serum amylase. A chest radiograph excluded free intraperitoneal gas. An ultrasound scan of her abdomen showed a thick-walled cyst in her pelvis of uncertain origin and a plain computed tomography showed that the cyst, with calcified structures, originated from her left ovary.On laparoscopic examination under general anesthesia, we found a cyst attached to her left ovary, which had twisted. After clipping and disconnecting its pedicle from her ovary, we attempted to aspirate its contents to facilitate extraction through a laparoscopic 10mm port. Because the contents of the cyst were too viscid, we could not achieve a reduction in cyst size through aspiration and the cyst was finally extracted through a 5cm transverse supra-pubic incision.Our patient recovered well after surgery and was discharged home on the first postoperative day, free of her generalized pruritus. Bisection of the cyst wall showed that it contained 200mL of thick, offensive milky brown fluid, hair and rudimentary teeth Figure . HistoloOur patient presented as an emergency due to torsion of the cyst on its long pedicle. In this regard, generalized pruritus likely arose from cyst wall ischemia and release of antigenic content into the general circulation, as has been demonstrated previously . BecauseWe searched MEDLINE using the key words ‘ovarian teratoma’, ‘torsion’, ‘pruritus’ and ‘itching’, and to the best of our knowledge believe that no similar previous case report exists.An unusual case of a woman with torsion of an ovarian teratoma presenting with abdominal pain and generalized pruritus, believed due to an antibody-mediated response, was resolved after surgical removal of the cyst. The cyst was found to contain sebaceous material, rudimentary teeth and hair.Written informed consent was obtained from the patient for publication of this case report and accompanying images. A copy of the written consent is available for review by the Editor-in-Chief of this journal.The authors declare that they have no competing interests.RD wrote and revised several drafts of the manuscript and photographed the cyst. AdS conceptualized the case report and reviewed the manuscript. RW performed the surgery and reviewed the manuscript. All authors read and approved the final manuscript.