TCF7L2 rs7903146 is associated with growth patterns from fetal life until infancy.An inverse association between birth weight and the risk of developing type 2 diabetes (T2D) in adulthood has been reported. This association may be explained by common genetic variants related to insulin secretion and resistance, since insulin is the most important growth factor in fetal life. The objective of this study was to examine whether T2D gene polymorphism TCF7L2 genotype was determined in 3,419 subjects in Generation R and in 566 subjects in the SGA cohort.This study was performed in two independent birth cohort studies, one prospective population-based (Generation R), and one of subjects born small-for-gestational-age (SGA cohort). Fetal growth was assessed by ultrasounds in second and third trimesters of pregnancy in Generation R. Growth in infancy was assessed in both cohorts at birth and at 6, 12 and 24 months postnatally. TCF7L2 genotype was also not associated with any pre- or postnatal growth characteristic in either Generation R or the SGA cohort.Minor allele frequency did not differ significantly (p = 0.47) between Generation R  and the SGA cohort . No differences at birth were found in gestational age or size  between the genotypes in either cohort. TCF7L2 genotype and fetal and early postnatal growth. Furthermore, this TCF7L2 polymorphism was not associated with an increased risk of SGA.We found no evidence for an association between  Several epidemiological studies have shown inverse associations between birth weight and metabolic diseases, including type 2 diabetes (T2D) in adulthood ,2. TheseTCF7L2 gene (rs7903146) increases the risk of T2D approximately 2-fold when two risk allele copies (TT) are present [TCF7L2 polymorphisms has been suggest to reduce proinsulin to insulin conversion [TCF7L2 gene have been shown to be associated with type 2 diabetes, although less strongly [TCF7L2 one of the most important candidate genes for explaining the associations between low birth weight and T2D.Genome-wide association (GWA) studies have consistently shown that the C>T substitution in ene rs790146 incre present . The T-astrongly , has beestrongly , though strongly . This postrongly . Such fiet al. were the first to investigate the association between TCF7L2 genotype and birth weight, and they found an association with maternal TCF7L2 genotype [TCF7L2 genotype did not influence fetal birth weight [TCF7L2 genotype and the risk of small size for gestational age [Freathy genotype . Each magenotype . After ah weight . This fih weight . In anotonal age . Birth wonal age . Furtheronal age ,19.TCF7L2 on growth than specific growth characteristic such as birth weight. We first assessed the associations of TCF7L2 rs7903146 with fetal and postnatal growth characteristics in a population-based prospective cohort study among 3,419 subjects followed from early fetal life onwards. Second, we assessed associations of this genotype with birth weight and postnatal growth in 566 small-for-gestational-age (SGA) children participating in an independent cohort study.Therefore we hypothesized that longitudinally measured fetal and postnatal growth are better parameters in the investigation of the possible effect of The Generation R Study is a population based prospective cohort study from early fetal life onwards. The study is designed to identify early environmental and genetic determinants of growth, development and health from fetal life until young adulthood. It has been described previously in detail ,21. Feta10 EFW = 1.5662 – 0.0108 (HC) + 0.0468 (AC) + 0.171 (FL) + 0.00034 (HC)2 – 0.003685 (AC * FL)) [Fetal ultrasound examinations were carried out during visits to one of the research centers. These fetal ultrasounds were used for establishing gestational age in the first trimester of pregnancy , as well as for assessing fetal growth characteristics in second  and third trimesters  of pregnancy . Fetal gC * FL)) . First tBirth weight, date of birth and gender were obtained from community midwife and hospital registries. Information on head circumference or length at birth was not available, but many children were measured during the first two months of life. Well-trained staff in community health centers obtained postnatal growth characteristics using standardized procedures. Based on the routine health care program, the visits at which these growth characteristics were measured were grouped into three age periods: 6 months (range 5 to 8.99); 12 months (range 9 to 12.99); and 24 months (range 23 to 34.99 months). Postnatally, head circumference was not measured at the age of 24 months.TCF7L2 genotyping and who also had Dutch or other Caucasian ethnicity as defined by having both parents born in the Netherlands or another European country   or at young adult age n = 252). Children were included in the SGA cohort when they were SGA at birth, had short stature (height standard deviation score (SDS) for age and gender of below – 2 2. Childr. Young aBirth characteristics of the SGA cohort were collected from hospital registries. The gestational age of the subjects was determined by ultrasound in the first trimester, if available, and otherwise calculated from the date of the last menstruation. Growth data  measured during the first two years of life were collected from records of hospitals, community health services and general practitioners. Longitudinal growth data were available in 272 participants in the SGA cohort Figure .TCF7L2 (rs7903146) gene was performed using Taqman allelic discrimination assay  and Abgene QPCR ROX mix . The genotyping reaction was amplified using the GeneAmp® PCR system 9600 (95°C (15 minutes), with 40 cycles of 94°C (15 seconds) and 60°C (1 minute)). The fluorescence was detected on the 7900HT Fast Real-Time PCR System (Applied Biosystems) and individual genotypes were determined using SDS software . Genotyping was successful in 98% and 91% of the samples in the Generation R and SGA cohort, respectively. To confirm the accuracy of the genotyping results, 276 randomly selected samples from the Generation R Study were genotyped for a second time with the same method. The error rate was less than 1%. The frequency distribution in Generation R did not deviate from the Hardy-Weinberg equilibrium in subjects with Dutch ethnicity nor did it deviate in the SGA cohort.DNA was collected from cord blood samples in the Generation R cohort and from peripheral venous blood samples in the SGA cohort. Cord blood for DNA isolation was available for 59% of all participating children of the Generation R cohort. When cord blood samples were missing, this result was mainly due to logistical constraints at the delivery. Venous blood samples were available in the complete SGA cohort. Genotyping of the C>T substitution in TCF7L2 genotype on growth, and allows for incomplete outcome data [With sample sizes in the Generation R Study of 3,419 and 2,675 subjects for fetal and postnatal analyses respectively, and assuming a statistical power level (1 – β) of 0.80, a level of significance (α) of 0.05 and a variance of 1.0, we were able to detect differences in growth characteristics of 0.048 SDS and 0.054 SDS respectively. First, differences in allele distribution between children born SGA (from the SGA cohort) and non-SGA subjects (from Generation R) were assessed. Differences were calculated using the Chi-square test. Second, we examined the differences in birth characteristics between genotype groups with linear regression analyses assuming an additive model. Weight, length and head circumference at birth and at different ages were analyzed using gender and age adjusted standard deviation scores (SDS) ,28. Stanome data . In thes0' reflects the intercept and the term including 'β1' reflects the slope of growth (weight or length) per week for the reference group (CC genotype). The terms including 'β2' and 'β3' reflect the age independent growth differences in weight (and length) between the different categories of the TCF7L2 genotype respectively [In this model, the term including 'βectively . All modectively . The occAll effect estimates are presented with their 95% confidence interval (95% CI). Statistical analyses were performed using the Statistical Analysis System version 9.1.3 , including the PROC MIXED module for unbalanced repeated measurements as well as the Statistical Package of Social Sciences version 15.0 for Windows .Subject characteristics of Generation R and SGA cohort are presented in Table No significant differences between genotype groups were observed in fetal growth characteristics in Generation R Table . No diffFinally, no differences were found in weight growth rate (SDS/year) from birth until the age of 2 years in either Generation R or the SGA cohort. Compared to the CC genotype, differences were -0.014 : -0.064, 0.036) SDS/year and -0.028  SDS/year, for the CT and TT genotype, respectively, in Generation R. In the SGA cohort, differences were -0.134  SDS/year and 0.002  SDS/year, for the CT and TT genotype, respectively, using the CC genotype as a reference. Similarly, no differences were found in height growth rate from birth to 2 years in either cohort (data not shown).In the current study, we found that T2D gene polymorphism TCF7L2 rs7903146 is not associated with growth in fetal life in the general population or with growth in early postnatal life in either the general population or in a cohort of subjects born SGA. We also confirmed previous suggestions that this variant of TCF7L2 is not associated with birth weight and, more importantly, demonstrated that it does not influence the fetal development using direct fetal measurements. Finally, we showed that this polymorphism does not appear to be associated with the risk of being born SGA.TCF7L2 with longitudinally measured growth patterns in fetal and early postnatal life in two independent birth cohorts. In the Generation R Study, DNA for genotyping was available in 59% of all subjects and was isolated from cord-blood. Missing cord-blood was mainly caused by logistical restraints at delivery. Children who were not genotyped had a shorter gestational age (p < 0.001) and were lighter at birth (p < 0.001) than subjects who were genotyped. Of all genotyped eligible subjects at baseline, 22% did not participate in follow-up measurements. In the SGA cohort, genotyping was successful in 91% of the subjects and longitudinally growth data were available in 48% of the cohort. Our effect estimates could be biased if the associations between genotypes and growth characteristics differed between those with and without postnatal growth data available. In the Generation R cohort, no differences were observed between children with and without postnatal growth measurements. In the SGA cohort the T-allele was slightly more frequent in subjects with postnatal growth measurements than in subjects without these measurements (p < 0.05). Finally, it could be possible that there is differential effect of genotype on growth according to availability of follow-up data. This bias would affect our estimates, though such a bias seems unlikely.To our knowledge, this study is the first to examine the association of CDKAL1 (rs7754840) and HHEX (rs1111875) genotype, and maternal TCF7L2 (rs7903146) genotype have been shown to affect birth weight. Pulizzi et al. demonstrated in the Helsinki Birth Cohort that fetal TCF7L2 genotype did not interact with birth weight to increase the risk of T2D in adulthood [TCF7L2 rs7903146 has been shown to have the strongest genetic effect on T2D and this result has been replicated in several studies [TCF7L2 is a very important candidate gene for explaining the association between low birth weight and T2D risk. Our study is the first to investigate the effect of TCF7L2 rs7903146 on longitudinal growth in early life. Longitudinal assessment of growth provides more information than just measurements at birth as we have demonstrated earlier that different fetal growth patterns may result in a similar birth weight [TCF7L2 rs7903146 influences fetal and postnatal growth, longitudinal growth data provide more information than birth weight alone.Several studies have investigated the effect of common genetic variants related to insulin action and secretion on early growth ,15,32,33h weight . Furtherh weight . Thus, tet al. found an increase of birth weight for each fetal and maternal risk allele [et al. found no effect of the fetal genotype of this polymorphism on birth weight. Since fetal and maternal genotypes are 50% correlated, it cannot be excluded that, when the risk allele is present in both mother and child, small effects of fetal genotype that reduce fetal growth could be masked by opposing effects of maternal genotype. Since maternal genotype was not available in our study, we were not able to test this hypothesis. However, we did not find any effect of fetal genotype on birth weight in the general population nor in a specific population of children with insufficient fetal growth resulting in small size for gestational age at birth. Our findings are therefore in line with the conclusions of these previous studies. Furthermore, we found no effect of fetal genotype on estimated fetal weight or weight during infancy, indicating that there is no evidence for any association between this fetal genotype and weight or change in weight during early life either. The effect of this polymorphism on the metabolic phenotype found in adults would therefore appear to develop after early childhood. Nonetheless, our results also could be explained by a lack of power and we cannot rule out that we were unable to detect smaller effects of this variant on early growth.Freathy k allele . They coTCF7L2 rs7903146 genotype on SGA. Cauchi et al. found no association between this genotype and SGA, using family-based association analyses in over 3,000 subjects of which 627 subjects were SGA [Regarding intra-uterine growth retardation, an earlier study examined the effect of were SGA . In thisTCF7L2 rs7903146 does not influence growth from early fetal life to infancy. Furthermore, minor allele frequency was not different in SGA subjects than in non-SGA subjects, indicating that it is unlikely that this polymorphism is associated with the risk of being born SGA. Systematic searches for common genetic variants by means of genome-wide association studies will enable us to obtain a more complete understanding of which genes are involved in growth in fetal life and infancy.In summary, our results suggest that The authors declare that they have no competing interests.DOMK, SWKdK, ACSHK and VWVJ made substantial contributions to conception and design of the manuscript and the analysis and interpretation of data. CMvD and AGU were involved in the genetic analysis of the data. AH and HAM were involved in the design of the cohort. EAPS was responsible for the prenatal growth data collection. All authors were involved in drafting and revising the manuscript and have given final approval of the version to be published.The pre-publication history for this paper can be accessed here:


An examination of birth weight in a Swedish cohort study of 38 566 women showed no significant association between birth weight and endometrial cancer, but supported a protective role for low birth weight for premenopausal breast cancer. Further, the role for low birth weight for breast cancer risk is not fully worked out. Adjustment for adult body mass index (BMI) is highly relevant, as it is associated with birth weight , emigration before the start of the follow-up (n=16), hysterectomy (n=1501), twins (n=1154), and with missing information on birth weight or adult BMI (n=7769). Thus, the final analysis was conducted on 38 566 women. The study was approved by the ethical committee at the University of Uppsala and the Karolinska Institute.The study cohort consisted of 49 259 Swedish women in the Women's Lifestyle and Health study . For all models, the attained age at follow-up was used as time scale and stratified by 5-year birth cohorts. Hazard ratios (HR) were declared statistically significant when the associated two-sided 95% Wald-type confidence interval did not cover the value one. This corresponds to a two-sided 5% level of significance. All statistical analyses were carried out using the SAS software version 9.1, .2), whereas 28% (20) were overweight (BMI 25–30 kg/m2), and 16% (12) were obese (⩾30 kg/m2). The corresponding figures for the breast cancer cases were: 76% (501) normal weight, 19% (123) overweight, and 5% (33) obese. When compared to the 38 566 women who were included in the analysis, the distribution of the baseline characteristics were similar for the 7769 women who were excluded from the study owing to missing values of birth weight or adult BMI (data not shown).The baseline characteristics of the 38 566 women in this study are presented in 2). Obese women (BMI⩾30 kg/m2) had a statistically significant increased risk of endometrial cancer  compared to normal weight women (BMI<25).The HR for endometrial and breast cancer risk for different birth weight categories are shown in The crude HR for breast cancer risk, when comparing women with low birth weight (<2.5 kg) to women with the highest birth weight (>3 kg) was 0.65 (95% CI: 0.43–0.99). The results were not affected by adjustment for adult BMI.The HR for endometrial and breast cancer reported in We found no evidence of an association between low birth weight, and reduced endometrial cancer risk, although our data supported earlier findings that a higher birth weight increases breast cancer risk. The results also suggest that low birth weight is protective for breast cancer. Adult BMI did not modify the effect of a low birth weight on endometrial or breast cancer risk.Strengths of this study are the prospective design, and the complete follow-up. One major limitation is that we only had three categories of birth weight, and the highest category included infants of normal as well as of high birth weight. Earlier studies have suggested that birth weights above 4000 g increase breast cancer risk (Another limitation of the study was that we could not adjust for gestational age, which has been shown to strengthen the positive association between birth weight and breast cancer risk (The inconsistency between this study and the only other study on endometrial cancer (In conclusion, no association between birth weight and endometrial cancer risk was found, whereas a low birth weight was suggested to reduce the risk of developing breast cancer among women born in the 1940–1960s.


Maternal exposure to wood fuel smoke may lead to impaired fetal growth due to hypoxia and or oxidative stress from smoke constituents such as carbon monoxide and particulate matter.We studied the risk of low birth weight (LBW) and reduced mean birth weight in relation to reported use of wood for cooking during the prenatal period, compared with natural gas (NG).We studied a historical cohort of women who had a singleton live birth in the years 2000–2002, from a semirural area of Pakistan. Infant’s birth weight was obtained from records, and prenatal records had data for maternal body mass index and parity. Cooking habits, daytime sleep habits, and type of fuel used during the pregnancies in 2000–2002 were ascertained by a survey done in 2004–2005. We performed multiple linear and logistic regression modeling using propensity scores to adjust for confounding variables.p < 0.06) in NG users. Infants born to wood users averaged 82 g lighter than infants born to NG users when weight was adjusted for confounders (p < 0.07). The rate of LBW  was 22.7% among wood users compared with 15.0% in NG users (p < 0.01), for an adjusted relative risk of 1.64 . The population attributable risk for LBW explained by wood use was estimated to be 24%.Unadjusted mean (± SD) birth weight was 2.78 ± 0.45 kg in wood users, and 2.84 ± 0.43 kg (Cooking with wood fuel during pregnancy, a potentially modifiable exposure, was associated with LBW and marginally lower mean birth weight compared with using NG. Low birth weight (LBW) is associated with high mortality and morbidity in infants . LBW leahuqqa) and chewing tobacco, however, is not uncommon in rural areas  in Karachi. The study area was Rehri Goth, a poor, semirural community located 50 km from Karachi. All of the defined area was included in the house-to-house survey when enrolling pregnant women, 2000–2002, and the same method was adopted in the survey in 2005. Traditional birth attendants were the main source of maternal care during pregnancy and labor.Rehri Goth is a coastal fishing village in Pakistan with a population of > 35,000 residents. More than 40% of the population is poor, using the definition of having an annual income of 625 rupees (Rs) per capita . The areAll married women 15–45 years of age who delivered a singleton live birth during 2000–2002 were included. Pregnancies were identified by interviewing women during house-to-house surveys in 2000–2002.A total of 1,111 women with at least one child < 5 years of age were identified in the area. A total of 817 of these women could be matched with birth record data at AKU  by usingn = 30; 15 wood and 15 natural gas (NG) users] who were not part of the cohort but had at least one child < 5 years of age. Informed, signed consent was obtained before every interview. Interviews were conducted in Urdu with all willing participants who could be located and who were willing to participate.The questionnaire was pilot tested in a sample of women .n = 6). Interview forms from the surveys done in 2000–2002 and 2004–2005 were checked daily for errors, and data entry was validated by double data entry. All interview forms were coded by the field supervisors before double data entry. After data entry, data were checked and errors corrected by a team of data management unit member, field supervisors, and principal investigator. All field workers were trained and certified in their data collection techniques before they were permitted to collect data in field.We assessed data from records for missing values, errors, and extreme values; extreme values for the upper 95th percentile and lower 5th percentile were verified from original record data. Extreme values of ≥ 60 cm were excluded for the crown–heel length of the neonate ; maternal age, body mass index (BMI), gravidity, and socioeconomic status were potential confounding variables. Propensity scores were developed to create probabilities of exposure to certain types of fuel (wood and NG). The propensity scores were calculated primarily to balance the groups and thus adjust simultaneously for many confounding factors while minimizing loss of effective sample size, which would have increased confidence intervals (CIs) if traditional logistic regression models had been used. Interpretation of models is facilitated with one-dimensional propensity scores, which accommodate many more variables than the traditional methods of adjustment. Participants with lower propensity scores were least likely to use wood, as determined by modeling for determinants of wood use. For instance, if a wood user and an NG user have the same propensity score, these two participants would have the same probability to use wood (if they could have the freedom to choose), and hence can be compared on the outcome measurements.Continuous variables were summarized with means and standard deviations and checked for normality of distributions. We compared the groups using Student’s p < 0.10 were retained in the model. We computed the predicted probability of wood use for each study participant based on the propensity scores. Propensity scores were included in multivariate models to control for differences in covariate distributions between the groups  were converted to relative risks (RRs) , and LevIn multiple linear regression models, birth weight and head circumference were regressed on fuel type, propensity scores, and covariates associated with the outcome. Multiple logistic regression modeling evaluated the relation of wood fuel use to high and low PI, with each compared with PI within one SD of the mean.Data were entered in Epinfo, version 6 , and analyses were conducted using statistical software SPSS version 12.00 for Windows .n = 506), cohort members living in the area (n = 806) had a more recent year of birth (p < 0.001), infants with smaller head circumference (p < 0.04), and early neonatal assessment (p < 0.04). The final sample totaled 634 participants, excluding those who reported changes or inconsistent use of single fuel type in the preceding 5 years (n = 94), did not cook food during the index pregnancy (n = 74), or used fuels other than wood or NG (n = 4) (Compared with those who left the area or refused to participate ( (n = 4) .n = 12), head circumference (n = 6), neonatal assessment day (n = 12), maternal gravidity status (n = 58), parity (n = 154), and BMI (n = 7). Compared with infants born to NG users, infants born to wood users had a lower mean birth weight (p < 0.06) and higher LBW rate (p < 0.05) (Record data were missing for crown–heel length (LBW rate . Infants < 0.05) . Compare < 0.05) .Compared with NG users, significantly more wood users had monthly household incomes less than the median (US$50/month) and lower-quality houses made of a mixture of straw and bricks . HousehoOverall, wood use was associated with LBW . Prenatal care by doctor had a higher relative risk of LBW compared with that provided by a traditional birth attendant ; this could be attributed to specific medical needs requiring medical attention during pregnancy. Low maternal BMI at the time of enrollment (2000–2002) and at the time of interview (2004–2005) were significantly associated with delivering an LBW infant.A large number of participants reported cooking frequency as more than twice per day and duration of cooking as 180 min (medians); and stratifying data on medians did not result in exactly 50% of data by fuel type. Less frequent cooking was associated with higher RR for LBW among wood  and NG users , and longer duration of cooking was significantly protective for LBW among NG users  but not among wood users . Compared with NG users, wood users had a significantly increased RR of LBW for greater frequency of cooking , longer duration of fuel burning , maternal duration of stay in the kitchen during fuel burning , and longer duration of cooking .2) maternal BMI (p < 0.001), maternal age < 25 years (p < 0.05), history of prenatal tobacco use (p < 0.03), maternal smoking (p < 0.08), wood use (p < 0.06), lower frequency of cooking per day (p < 0.08), higher crowding index (p < 0.06), infant assessment within 48 hr of birth and birth year before 2002 were related to lower mean birth weight. Mean birth weights tended to increase from 2000 to 2002 (p < 0.07). Lower crown–heel length of the infant was associated with younger maternal age (p < 0.005), less time spent cooking (p < 0.03), lower gravidity (p < 0.01), lower maternal BMI (p < 0.06), female infant (p < 0.05), infant assessment within 48 hr of birth (p < 0.001), and infants born after 2000 (p < 0.002). Lower head circumference was associated with paternal illiteracy (p < 0.07), assessment after 48 hr after birth (p < 0.002), being born after 2000 (p < 0.002), and higher frequency of cooking per day (p < 0.06). Lower PI was significantly associated with maternal tobacco use in pregnancy (p < 0.006) and births before 2002 (p < 0.002), and insignificantly with maternal illiteracy (p < 0.11). Female infants had lower mean birth weight and length, compared with male infants, but head circumference and PI did not differ by sex.Unadjusted analyses showed that low . Among wood users whose infants were LBW, 52.2% (22 of 42) had lower PI (p < 0.08), compared with 30% (7 of 23) in NG users.NG users had 11.2% (23 of 204) and wood users had 15.5% (42 of 270) of their infants with a lower PI  in a linear regression model, unadjusted for propensity scores. However, a nonsignificant difference was observed when the propensity score was added to the model   .The regression models showed that a later assessment of birth weight was associated with higher birth weight, and prenatal care was related to LBW. In this study we found that exposure to wood fuel burning for cooking during pregnancy was associated with a significantly increased odds of LBW (OR = 1.86) and a reduced mean birth weight of infants (by 82 g), adjusting for potential covariates. More frequent cooking was associated with a higher mean birth weight, possibly because the families that were better off economically could afford to cook more times per day. But increased time spent in the kitchen during fuel burning for cooking was associated with a higher risk of LBW in wood users, though not in NG users. The contribution of wood smoke pollutants may not appear to be substantial in the presence of established nutritional, reproductive, lifestyle, and socioeconomic risk factors, but fuel used for cooking is potentially much more modifiable than these other factors. The use of wood fuel for cooking is estimated to be approximately 53% in Pakistan, and the overall use of biomass fuels is > 70%. The population-attributable risk of 24% for LBW indicates that removal of wood fuel exposure from the population has the potential to reduce the risk of LBW from the current 19% rate to 14.4%, which is just below the targeted rate recommended by the Our model for mean birth weight included many confounding factors; thus, it is likely that the association of mean birth weight with wood use would have been statistically significant if we had had a larger sample. The results of this study are consistent for the negative effect of use of wood fuel on birth weight with the two other studies that have evaluated the role of wood fuel in reducing mean birth weight . A reducp < 0.030). Another factor that could reduce head circumference is day of assessment, because initial assessment may be affected by edema or hematoma after birth, though day of assessment did not differ significantly by self-reported maternal smoking during pregnancy. The finding that lower head circumference was associated with increased frequency of cooking and year of birth was somewhat inconsistent with our finding that birth weight increased with increased frequency of cooking (p < 0.08) and with later (more recent) year of birth (p < 0.07).Our model for low PI (indicator of infant wasting) showed insignificant associations with use of wood fuel, although more wood users had infants who had a low PI (disproportionate in weight for the given height) than did NG users. It is likely that wood smoke could affect body proportions, as has been reported for cigarette smoking , but thep < 0.001) on birth weight; this relationship was linear over a longer period of time, and the correlation was statistically significant both parametrically and nonparametrically. However, in the linear regression model increasing birth weight occurred with increasing day of assessment (which occurred more in wood users). Our study sample also included only participants who had live singleton births; births that ended in miscarriages or fetal or early neonatal deaths were excluded. Considering these observations and the high proportion of male infants who had LBW than female infants among NG users, if any bias occurred it was toward the null, thus making our estimates of effects of use of wood fuel conservative.The study had several limitations. First, we did not have complete data for gestational age on most pregnancies to enable us to control for it and to evaluate the proportions of preterm births and infants with intrauterine growth retardation in the two fuel groups. Second, we were missing data on some covariates, such as use of prenatal multivitamins or iron supplements and any prenatal illnesses, which could have led to some uncontrolled confounding. The women in our study were followed until the end of pregnancy and were referred by health workers for routine prenatal supplements, including iron and folic acid; in many cases, health workers provided supplements at home, regardless of fuel type. We ascertained and analyzed self-reported anemia, diabetes, hypertension, and other illnesses but, possibly because of inaccuracy of recall, we did not find any important relationships between them and LBW or mean birth weight. Third, because the day of assessment of newborns was later for wood users than for NG users, it is possible that some of the LBW in the wood fuel group was attributed to infants losing weight after birth, as is normal, and having their weight measured later, although the association persisted after adjustments for day of neonatal assessment. The day of measurement of birth weight had an influence , and socioeconomic factors, thus minimizing the likelihood of uncontrolled confounding affecting the results. Use of wood fuel and LBW are linked with poverty , and povThis large population-based study adds to the growing body of evidence concerning the effects of wood use on risk of LBW and corroborates the known associations of maternal nutrition, reproductive, and socioeconomic factors with infant birth weight. Notably, accounting for differences in demographic, socioeconomic, and nutrition factors further strengthened the validity of our findings that associated prenatal wood fuel use with increased risk of LBW. Thus, preventive measures designed to minimize or remove this exposure could dramatically reduce the occurrence of LBW and thus the morbidity and mortality that results from LBW.


P for trend=0.02). Birth weights in the highest quartile (3730 g or more) were associated with 40% higher risk  of breast cancer compared to birth weights in the lowest quartile (less than 3090 g). For birth length, the odds ratio for women who were 51.5 cm or more (highest quartile) was 1.3  compared to being less than 50 cm (lowest quartile) at birth. Adjustment for age at first birth and parity did not change these estimates. Placenta weight was not associated with breast cancer risk. This study provides strong evidence that intrauterine factors may influence future risk of breast cancer. A common feature of such factors would be their ability to stimulate foetal growth and, simultaneously, to influence intrauterine development of the mammary gland.The hypothesis that birth weight is positively associated with adult risk of breast cancer implies that factors related to intrauterine growth may be important for the development of this malignancy. Using stored birth records from the two main hospitals in Trondheim and Bergen, Norway, we collected information on birth weight, birth length and placenta weight among 373 women who developed breast cancer. From the same archives, we selected as controls 1150 women of identical age as the cases without a history of breast cancer. Information on age at first birth and parity were collected from the Central Person Registry in Norway. Based on conditional logistic regression analysis, breast cancer risk was positively associated with birth weight and with birth length  86, 89–91. DOI: 10.1038/sj/bjc/6600011www.bjcancer.comThe Cancer Research Campaign© 2002  These potential controls were born consecutively to the case in the same city and were residents of Trondheim or Bergen at the time of the case's diagnosis.The population base of this study comprised all female residents of Trondheim and Bergen, Norway, who were born at the two main hospitals in those cities between 1910 and 1970. In order to identify eligible participants, we first used the Norwegian Cancer Registry to identify 1035 breast cancer cases who were residents in these cities at the time of diagnosis. The Central Person Registry, a department of Statistics Norway, provided the necessary information to ascertain that 719 of these cases were also born in Trondheim or Bergen. This registry maintains continuously updated records on each woman's residential and childbearing history, including her mother's identity. For each case of breast cancer, we used the Central Person Registry to select four women . Due to limited resources, we aimed at identifying three controls per case in Bergen, and found 481 controls (2.6 per case). Thus, birth records were identified for 1150 of the 2876 potential control women who were born at the two main hospitals , birth length (centimetres), and placenta weight (grams). We also collected maternal information about height, marital status, and father's occupation; which was available for nearly half of the participants. The information on childbearing history  of the cases and controls was collected from the Central Person Registry in Norway. We used conditional logistic regression to examine the effect of birth weight, length, and placenta weight on breast cancer risk, categorizing each study factor by quartiles, based on the distribution among controls, and making adjustment for age at first birth and parity.P for trend=0.02). Birth weight in the highest quartile (⩾3730 g) was associated with an odds ratio of 1.4  compared to the lowest quartile (⩾3090 g). For birth length, there was a similar positive association (P for trend=0.02), with an odds ratio of 1.3 , for the highest (⩾51.5 cm) versus the lowest quartile (<50 cm). Adjustment for indicators of the mother's socio-economic status  did not materially change these estimates in the subset of subjects for whom this information was available (data not shown). There was no association between placenta weight and breast cancer risk (As expected, increasing age at first birth was associated with increasing risk of breast cancer, and there was a reduction in risk with increasing parity (Table 2Studies of the effect of birth weight on breast cancer risk have yielded different results  employedBirth length is typically measured by the half centimetre, resulting in a narrow variation compared to that of birth weight, which is measured in grams. Nonetheless, we found a positive association between birth length and breast cancer risk, and the strength of the association was fairly similar to the one related to birth weight. Previous studies with information on birth length have found only weak, and not statistically significant associations between birth length and breast cancer risk (We found no evidence for any association between placenta weight and breast cancer. Previously, placenta weight has only been examined in the Swedish studies  that alsThe positive association observed between birth size and breast cancer in the present study could be confounded by socio-economic factors. We did have some information that could indicate socio-economic differences, such as father's occupation, the mother's marital status at birth, and mother's height. The latter could also be correlated with the offspring's birth size for genetic reasons. Due to missing data on these factors for a substantial proportion of the participants we could not resolve this issue. Nonetheless, we did explore the potential confounding by these factors among participants for whom this information was available, but the main results were not substantially altered.This study provides strong evidence that intrauterine factors may influence future risk of breast cancer. Since positive associations were found for birth weight and birth length, the relevant underlying factors would most likely be linked to fetal growth. A common feature of such factors would therefore be their ability to stimulate fetal growth and, simultaneously, to influence intrauterine development of the mammary gland.


Background. Magnetic Resonance (MR) diffusiontensor imaging (DTI) is able to quantify in vivo tissuemicrostructure properties and to detect disease related pathologyof the central nervous system. Nevertheless, DTI is limited by lowspatial resolution associated with its low signal-to-noise-ratio(SNR). Aim. The aim is to select a DTI sequencefor brain clinical studies, optimizing SNR and resolution. Methods and Results. We applied 6 methods for SNRcomputation in 26 DTI sequences with different parameters using 4healthy volunteers (HV). We choosed two DTI sequences for theirhigh SNR, they differed by voxel size and b-value. Subsequently,the two selected sequences were acquired from 30 multiplesclerosis (MS) patients with different disability and lesion loadand 18 age matched HV. We observed high concordance between meandiffusivity (MD) and fractional anysotropy (FA), nonetheless theDTI sequence with smaller voxel size displayed a bettercorrelation with disease progression, despite a slightly lowerSNR. The reliability of corpus callosum (CC) fiber tracking withthe chosen DTI sequences was also tested. Conclusions. The sensitivity of DTI-derivedindices to MS-related tissue abnormalities indicates that theoptimized sequence may be a powerful tool in studies aimed atmonitoring the disease course and severity. Magnetic Resonance (MR) diffusion tensor imaging (DTI) allows in vivo examination of the tissue microstructure, obtained by exploiting the properties of water diffusion. The DT computed for each voxel allowed us to calculate the magnitude of water diffusion, reflected by the mean diffusivity (MD) and the degree of anisotropy, which is a measure of tissue organization, expressed as an a-dimensional index, such as fractional anisotropy (FA) . The patThe best acquisition and postprocessing strategies for DTI sequences in the disease, especially in MS, are still a matter of debate , 6, 7.b = 0 s/mm2 images, this technique is vulnerable to poor SNR values: the background noise level close to the low diffusion weighted signal would overestimate the signal itself and consequently underestimate the magnitude of diffusion. The SNR of the b = 0 s/mm2 images should be at least 20 to obtain unbiased DTI-derived measures. Many methods for SNR evaluation in MR images are available and they differ for the estimation of the noise variance. They are commonly subdivided into two classes: single magnitude image methods derive the noise from a large, uniform background region : , paying μair) is equal to the SD of the noise, multiplied for the coefficient The standard deviation of noise was estimated from a ROI of 20 × 20 = 400 pixels, extracted from background (air). Since MR noise in the air follows Rayleigh distribution, the mean value of the signal in the second ROI  or vice versa.This method was similar to the method 2. We considered two images (A and B) obtained from two subsequent acquisitions of the same slice. The signal was the mean value of the pixels in a ROI on the first image (A). Then, we considered a second ROI on the second image (B), located as the first ROI in the first image. The SD of the noise was evaluated in the same ROI position and computed as suggested by Ogura et al.  with 5)5):(5)σ=m) value at which the background histogram attains a maximum (mmax air): noise SD was estimated as the mode of the Probability Density Function histogram [Since MRI noise in the air follows Rayleigh distribution, the noise variance can be estimated by searching for the magnitude -weighted images, using the corresponding T2-weighted images to increase confidence in lesion identification. Then, lesion volume (ml) was calculated and segmented lesions were used for masking DTI see , using Jhttp://www.fil.ion.ucl.ac.uk/spm/) and maximum image in-homogeneity correction [3D-T1 MP-RAGE images were automatically segmented to GM, WM and cerebrospinal fluid (CSF), using SPM5  based nonlinear transformation. Then diffusion gradient directions were corrected for scanner settings  and diffusion tensor was determined for each voxel using the freely available Diffusion Toolkit software, version 0.4.2 (http://www.trackvis.org/) with linear least-squares fitting method [DTI data were corrected for eddy-current distortion by FSL package, which registered the 12 diffusion-weighted volumes to the g method . The tenROIs of lesions individuated on T2-images were masked out from MD and FA maps, in order to estimate NAWM damage.GM and WM mutual exclusive masks were superimposed to MD and FA maps, and the corresponding histograms were produced. The erosion of the first-line outer voxels from the mutual exclusive masks excluded the contribution of partial volume effect from the surrounding CSF to the observed GM and WM diffusivity changes and WM anisotropy changes. Average MD was computed for GM and NAWM. Average FA was derived only for the NAWM, since no preferential direction of water molecular motion is expected to occur in the GM, due to the absence of a microstructural anisotropic organization of this tissue compartment.http://www.trackvis.org/) and visualized by the freely available software TrackVis v0.4.2 (http://www.trackvis.org/). The brute force approach and deterministic streamline-based fiber tracking were used, with FA-map as masking image and angle termination of 35°. For track selection, the one-ROI approach was used: CC was identified and segmented in the three mid-sagittal adjacent slices of FA-map [The reliability of fiber tracking with the 2 DTI sequences was tested using Diffusion Toolkit v0.4.2 (f FA-map . FA and MD histograms were derived for CC fiber tracts (CC-FA and CC-MD).A graphical display allowed to compare the six methods of SNR estimation and the quality of the sequences in terms of SNR.We estimate the intraclass-correlation coefficients between the 2 DTI sequences used in the study, regarding the values of NAWM-FA, NAWM-MD, and GM-MD of all the 48 subjects (HV and MS patients). Spearman's correlation coefficient (SCC) was assessed to estimate the correlation between DTI-derived measures  and the subjects' condition .As expected, the six SNR evaluation methods gave different absolute numerical values. Nevertheless, the changes through slices  and throb-value, TE and TR: this kind of graphical representation showed clearly the increase of SNR with the increase of the voxel size. A similar representation was done for sequences with the same parameters but the b-value, giving the result of SNR decreasing with the increasing of the diffusion-sensitivity coefficient, in particular the SNR estimated on images obtained from sequences with b-value of 1500 s/mm2 was 20% less than the SNR of sequences with b-value of 1000 s/mm2. The same analysis confirmed that the minimum TE feasible for the MR-scanner had to be selected, as expected, since DTI is T2 weighted.SNRs were plotted for sequences ordered by ascending voxel size and with the same The sequence with the highest SNR by all methods was DTI-A, which is characterized by parameters in the range recommended by Pagani et al.  for mult2. DTI-B SNR is lower than DTI-A SNR, less than 15%.Another sequence (DTI-B) was selected for the high SNR between those of pixel size of about 1 × 1 mmThe SNR comparison of the two selected sequences is shown in Figures The intraclass-correlation coefficients ranged from 0.91 to 0.99, showing high concordance of the parameters derived from DTI-A and DTI-B .P < .01) and subjects' condition as shown in The SCC showed that both DTI sequences separated HV from RRMS and SPMS patients, but that SCCs between DTI-B were higher than those between DTI-A ((i) Tractography algorithm was obtained with both the selected DTI sequences for all HV (in (ii) Tractography algorithm was obtained with both the selected DTI sequences for 28 of the 30 MS patients  but failIn this study we improved the quality of DTI sequences, looking for a compromise between SNR and spatial resolution. SNR values computed with different methods showed different bias and sensitivity to the noise level: this observation has to be further investigated. Despite that, at the aim of the present work, all methods were in accordance with the whole data set in pointing sequences DTI-A and DTI-B as the best ones without exception (SNR DTI-A > SNR DTI-B). These concordant evaluations allowed us to produce an automatic DTI sequences quality evaluation and to preliminary select two DTI sequences among 26. The two selected sequences had the best trade-off between SNR, voxel size, and diffusion sensing. Even if DTI-B has a lower SNR compared to DTI-A, the loss of maximum 15% in SNR was compensated by a higher resolution, which is a key element in determining tractographic reconstruction quality . Both DTThe optimum result is the production of CC individual-based tractography in 28 of 30 patients, with fiber tracts reconstructed even if they passed through a lesion. Both focal and diffuse alterations of tissue organization, which result in a decreased anisotropy and a consequent increase in uncertainty of the primary eigenvector of the DTI, are the well-known cause of the failure of tractography in MS in the previous studies , the numIn order to increase SNR, more than one average is usually acquired, but too many averages amplify coregistration errors and raise acquisition time and subject movements. In our DTI protocol we choose to acquire 2 averages (runs) for every diffusion sequence. In The noise, estimated with different methods, is almost constant over the slices : for exaBesides SNR examinations, even resolution has to be considered in DTI sequence parameters selection. Indeed, FA and MD are also influenced by the voxel size, due to the increment of the radial eigenvalues in a large voxel . FurtherFor the above reasons we included also DTI-B in the clinical protocol, due to the smaller voxel size, even if DTI-A had the higher SNR.Accurate FA and MD estimations improve the reliability of tractography, which is prone to errors: some of them are subjective  and some are intrinsic in the DTI sequence used. Indeed, bias in the estimation of diffusion tensor eigenvectors and eigenvalues damaged fiber tracking because it causes false or missing fibers , 30. SevThe results about SNR computed with different methods Figures and 4 shob-value (DTI-B) achieved a better correlation with the presence of MS disease. Even if DTI-B sequence has less slices than DTI-A, it covered the entire CC tracts due to the acquired slab position. Appropriate positioning of the acquisition slab should be evaluated in further studies in order to analyze other fiber bundles. Both our selected DTI sequences were able to quantify a tissue damage in MS, leading to distinguish between MS patients and HV and between the different MS phenotypes. However, the sequence with higher resolution and higher Finally, the proposed sequence and procedure showed higher reliability for fiber tracking and were able to discriminate the presence of MS disease even when severe lesional patterns were observed and may therefore be considered a potential powerful tool for studies to monitor the disease course and severity.


In the most recent revision (2003) of the U.S. standard certificate of live births, the National Center for Health Statistics recommended that all states collect maternal and paternal usual occupation. Because such information might be useful in the surveillance of job-related risk areas, we assessed the quality of parental work information on the U.S. birth certificate.Occupational histories obtained from maternal interviews with Texas (USA) participants in the National Birth Defects Prevention Study were linked to and compared with parental work information on birth certificates. With occupational information from interviews serving as the gold standard, we assessed the quality of occupational information on the birth certificate with measures of sensitivity, specificity, and the kappa statistic.Of the 649 births available for study, parental occupation agreed between the birth certificate and interview for 77% of mothers and 63% of fathers with similar agreement by case-control status. Among occupations and industries with 10 or more workers by interview, sensitivity of the birth certificate information ranged from 35% to 100% for occupational groups and 55% to 100% for industrial sectors. Specificities of occupations/industries studied ranged from 93 to 100%. Kappa statistics for maternal occupations (0.76 to 0.90) and industries (0.59 to 0.94) were higher than those for paternal occupations (0.48 to 0.92) and industries (0.47 to 0.89). Mothers were frequently misclassified as homemakers or otherwise unemployed while the paternal information was often missing altogether on the birth certificate. Women who worked as health diagnosing and treating practitioners were the least likely (0%) and women in food preparation or serving occupations were the most likely (65%) to be misclassified as not employed on the birth certificate. Among fathers, the proportion of missing occupations was the lowest for occupations in business or financial operations (0%) and highest for occupations in food preparation and serving (30%).Sensitivity of occupation/industry information on birth certificates varies although the specificity of such information may exceed 95%. Quality of this information also varies by maternal and paternal occupation with misclassification as homemaker a limiting factor among maternal and missing information a limiting factor among paternal work information. As part of the National Occupational Research Agenda (NORA) for reproductive health research, the U.S. National Institute for Occupational Safety and Health (NIOSH) recommended increased surveillance activities including the evaluation of exposure data from existing surveillance systems and expansion of birth defects surveillance systems . NumbersIn the most recent revision 2003) of the U.S. standard certificate of live births, the National Center for Health Statistics (NCHS) recommended the collection of maternal and paternal usual occupation (worked during the last year) if states could fund the collection and coding of these data  of the U. A surveIn this study, we evaluated parental work information recorded on the birth certificate in which the current methods of job and industry classification were used. We compared the maternal and paternal occupation recorded on the birth certificate with information obtained from telephone interviews of mothers of births with and without major malformations. We also examined the degree of misclassification as unemployed  and the proportions of missing values on the birth certificate across parental occupations and industrial sectors of work.The study population consisted of Texas participants in the National Birth Defects Prevention Study (NBDPS) who gave birth to offspring with selected congenital defects or births without major birth defects during 1997–2000. The NBDPS is a population-based, case-control study that has been ongoing since 1997 and includes study populations in 10 states ,13. DuriIn the NBDPS, interviews are targeted for completion within 6 months of delivery with the maximum time from delivery/termination to interview of no more than 24 months . A higheJob titles and industries on the NBDPS questionnaires were coded by the Centers for Disease Control and Prevention according to the SOC and NAICS systems ,11. We aWe examined whether occupational and industrial codes between the birth certificate and interview agreed for each Texas study participant. If the codes differed, we also compared job titles and industrial names and descriptions between the two sources. Some mothers and fathers had more than one job during the exposure windows of interest; jobs and industries were considered in agreement if at least one job or industry identified in the interview was in agreement with that listed on the subject's birth certificate. All assignments were made without knowledge of case or control status. In addition to examining overall agreement of the two sources by job title and place of work, percentages of agreement were examined by case/control status, maternal age at delivery, maternal education and race/ethnicity, and periconceptional folic acid use. Percentages of agreement were calculated with missing values included and excluded. The chi-square statistic or Fisher exact test (for any table with one or more cells with an expected frequency of less than 5) were used to assess the chance likelihood of observed variability .For both data sources, job titles were grouped into the SOC major groups except community and social services occupations, health care practitioner and technical occupations, and construction and extraction occupations which were divided into finer categories. Industry of work was grouped by NAICS sector with the exception of wholesale and retail trade which were combined. With the interview information considered as the "gold standard", we calculated sensitivity (number of workers classified as working in a given category according to both birth certificate and interview divided by the number of workers identified as working in the respective category by interview), specificity (number of workers classified as not working in a given category according to both birth certificate and interview divided by the number of workers identified as not working in the respective category by interview), and 95% confidence intervals for each measure with statistical programs developed by Abramson and Gahlinger . To asseWe also examined the distributions of missing values and misclassification as unemployed  for maternal and paternal occupations and industrial sectors. For maternal occupations and industrial sectors, "students" were combined with "homemakers" because of the small percentage (2.9%) of women misclassified as students on the birth certificate.A total of 307 case-births with neural tube defects, conotruncal heart defects, or oral clefts and 342 control births among Texas participants of the NBDPS were linked to their live birth certificates (1997–2000 births). This sample represented 100% of the Texas 1997–2000 controls, 88% of the conotruncal heart cases, 88% of the oral cleft cases, and 58% of the neural tube defect cases. Including participants with missing information, 77% of the maternal occupations and 63% of the paternal occupations agreed between the data sources. Excluding participants with missing information, 78% of the maternal occupations and 75% of the paternal occupations agreed. Table Because cases and controls were similar with respect to agreement of the birth certificate information with interviews, we combined these groups in analyses of comparisons of agreement by selected maternal characteristics and specific occupations and industries. Table For all maternal and paternal occupational groups studied, the specificity of the birth certificate was 95% or better indicating that if a parent was not working within a specific occupational group according to the interview, the birth certificate most of the time identified the parents as not working in the given group. For maternal occupations with 10 or more workers by interview, the percent of women misclassified as unemployed on the birth certificate varied considerably across occupational groups, ranging from 0 percent for women working as health diagnosing and treating practitioners to 65 percent for women working in food preparation and serving occupations Table . Among eAmong the paternal occupational groups with 10 or more workers by interview, sensitivity of the birth certificate ranged from 35% for health care support occupations to 100% for health diagnosing and treating practitioners and for military-specific occupations Table . Kappa sAmong industrial sectors with 10 or more workers by interview, specificities of industrial sectors on the birth certificate ranged from 93.4 to 99.5%, and sensitivities ranged from 55 to 100%. Except for the administration and support sector, kappas for maternal and paternal industrial sectors fell within the substantial or almost perfect categories of agreement. Sensitivities, specificities, and kappas for industrial sectors are shown in Tables The findings of this study indicate that parental occupation on the birth certificate may have some utility in the surveillance of occupational hazards in the workplace. Restricting comparisons to mothers who were gainfully employed according to the birth certificate and by interview, maternal occupation on these records exhibited excellent specificity and agreement (as measured by the kappa statistic) for most SOC occupational groups in which sufficient numbers of workers were available for analysis. We noted less agreement of the birth certificate with interview information regarding paternal occupation and maternal and paternal place of work than for maternal occupation. The limitations in using the birth certificate for occupational surveillance may differ for maternal and paternal occupations. In this study, mothers were frequently misclassified as homemakers or otherwise unemployed while the fathers' work information was often missing altogether. Furthermore, the problems of misclassification and missing values varied across occupational and industrial groups.Because some parents change occupations between conception and delivery, the designation of parental occupation on the birth certificate as "usual occupation" limits the utility of using these records in the surveillance of reproductive hazards having effects early in pregnancy. Among mothers and fathers in this study population who held more than one position during the index pregnancy, the job held at delivery usually agreed between the birth certificate and interview.As in other studies, we found overreporting of "homemaker" on the birth certificate ,7. This The findings in this study were similar to those in New York  and CaliIn the Santa Clara County study, the birth certificate was also examined for accuracy of paternal occupation for the period of three months prior to pregnancy and was noted to be comparable to interview information for approximately 80% of the records. In contrast, the birth certificate agreed for paternal occupation in only 63% of the records in the present study (with the missing cases included) in which we also included the first trimester as part of the exposure window. Part of this discrepancy might be explained by the higher proportion of unknown occupations (16%) in this study compared with the Santa Clara county study (6%). Using interview information, we examined types of occupations represented in the unknown category for birth certificates and noted a disproportionate number of occupations in food service, cleaning, construction, and the military according to the interview.Based on finding similar misclassification of occupation on the birth certificate by case control status in previous studies, investigators ,8 concluFurthermore, the proportion of missing paternal occupations on the birth certificate varied by reported maternal folic acid use during the periconceptional period. This differential in missing information by folic acid use could introduce bias in studies of paternal occupation and neural tube defects, oral clefts, and other defects for which insufficient folic acid is a risk factor. These findings underscore the need for caution when interpreting associations of parental occupation and birth defects based on birth certificate information. Choice of referent group might lead to overestimates as well as underestimates of associations if missing occupations vary by case-control status.Previous studies did not specifically examine the proportion of missing values on the birth certificate by parental occupation/industry. In the present study, paternal professional or managerial occupations that require more formal education were less likely to have missing information on the birth certificate than occupations requiring less education such as construction or food service occupations. Very few birth certificates were missing information regarding maternal occupation or industrial sector, but, as already discussed, mothers were frequently misclassified as homemakers.This study had several limitations including use of the maternal interview as the gold standard to evaluate the quality of paternal occupational information on the birth certificate. In telephone interviews, mothers provided information about fathers' occupations and places of work. Schnitzer et al.  found moWe also did not evaluate the accuracy of parental occupation in the fetal death records because this information was not available in the computerized fetal death record files in Texas. Although this exclusion amounted to only 20 cases in this study, it would be important to include fetal death records for occupational disease surveillance for several types of birth defects such as anencephaly in which fetal deaths represent a high proportion of cases.In conclusion, linkage of birth defect registry cases with their respective birth certificates and parental occupation shows some promise in surveillance of reproductive hazards in the workplace. The findings in this study also indicate that the quality of this information varies by parental status and by occupation. Misclassification of mothers as homemaker or otherwise unemployed on the birth certificate and missing information for paternal occupation may be problematic. The overall agreement between the birth certificate and maternal interview, degree of misclassification, and proportion of missing values also vary by occupations and industrial sectors. Because of these limitations, occupational surveillance findings based on the birth certificate should be followed up with studies based on parental interview and industrial hygiene assessments of workplace exposures.The author(s) declare that they have no competing interests.JDB conceived of and designed the study, carried out the data analyses, and wrote the first draft. LS and PHL participated in the design of the study, acquired the relevant data, assisted in the interpretation of results, and provided revisions to the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


Active pre-mRNA splicing occurs co-transcriptionally, and takes place throughout the nucleoplasm of eukaryotic cells. Splicing decisions are controlled by networks of nuclear RNA-binding proteins and their target sequences, sometimes in response to signalling pathways. Sam68 (Src-associated in mitosis 68 kDa) is the prototypic member of the STAR  family of RNA-binding proteins, which regulate splicing in response to signalling cascades. Nuclear Sam68 protein is concentrated within subnuclear organelles called SLM/Sam68 Nuclear Bodies (SNBs), which also contain some other splicing regulators, signalling components and nucleic acids.CD44 variable exon v5 and TJP1 exon 20 minigenes, these proteins did not, however, co-operate with each other in splicing regulation of these target exons.We used proteomics to search for the major interacting protein partners of nuclear Sam68. In addition to Sam68 itself and known Sam68-associated proteins , we identified hnRNP L as a novel Sam68-interacting protein partner. hnRNP L protein was predominantly present within small nuclear protein complexes approximating to the expected size of monomers and dimers, and was quantitatively associated with nucleic acids. hnRNP L spatially co-localised with Sam68 as a novel component of SNBs and was also observed within the general nucleoplasm. Localisation within SNBs was highly specific to hnRNP L and was not shared by the closely-related hnRNP LL protein, nor any of the other Sam68-interacting proteins we identified by proteomics. The interaction between Sam68 and hnRNP L proteins was observed in a cell line which exhibits low frequency of SNBs suggesting that this association also takes place outside SNBs. Although ectopic expression of hnRNP L and Sam68 proteins independently affected splicing of Here we identify hnRNP L as a novel SNB component. We show that, compared with other identified Sam68-associated hnRNP proteins and hnRNP LL, this co-localisation within SNBs is specific to hnRNP L. Our data suggest that the novel Sam68-hnRNP L protein interaction may have a distinct role within SNBs. Alternative splicing is regulated in part by a network of signalling pathways which respond to extracellular stimuli . One molCD44 variable exon v5 in response to Ras activation [CD44 exon v5 splicing by Sam68 involves interactions with U2AF65 to facilitate v5 exon definition [Sam68 has been reported to be associated with a number of different proteins involved in RNA processing, transcription, and cell signalling. Sam68 induces ERK -mediated inclusion of tivation . Activatfinition  and nuclfinition ,10. The finition ,12.In cancer cells, Sam68 protein exhibits a general nucleoplasmic distribution but is also concentrated within subnuclear structures called SLM/Sam68 Nuclear Bodies (SNBs) . AlthougSam68 protein is expressed at high levels in the nuclei of prostate cancer cells ,14. We iIn addition to the three previously known Sam68-associated proteins, we also identified hnRNP L as a novel potential Sam68-interacting protein partner Figure . By CoomTo further verify the novel Sam68-hnRNP L detected interaction, HEK293 cells were transiently transfected with expression constructs encoding FLAG-tagged Sam68 and GST-tagged hnRNP L proteins. Anti-FLAG M2 agarose was used to immunoprecipitate FLAG-tagged proteins from HEK293 cells, and co-immunoprecipitated GST fusion proteins were detected using antisera specific to GST. Consistent with the observed interactions of endogenous proteins, hnRNP L-GST was efficiently co-immunoprecipitated with Sam68-FLAG , which also regulates signal-dependent splicing. We carried out further experiments to test if hnRNP LL protein would be also found within SNBs. Surprisingly, these experiments indicated that hnRNP LL protein exhibits a nucleoplasmic distribution outside SNBs Figure . The obsTo verify these localization findings, HeLa cells were transiently transfected with expression vectors encoding hnRNP L-GST, hnRNP LL-GST, or GST alone. We then detected the localisation of the ectopically expressed proteins using antisera to detect the GST tag, and compared this distribution with that of endogenous Sam68 protein Figure . ConsistThe frequency of cells in a population containing SNBs is cell line-dependent. To test whether the association of Sam68 and hnRNP L was dependent on the presence of SNBs, we performed side-by-side immunoprecipitations in whole cell extracts obtained from LNCaP and NIH3T3 cells (only ~5.5% of NIH3T3 cells have SNBs) . Sam68 pThe above experiments indicate that Sam68 and hnRNP L proteins interact and are both present within SNBs. Although SNBs contain some splicing regulators and RNA, they do not detectably contain key spliceosome components . Hence SCD44 variable exon v5 [TJP1 exon 20 inclusion is repressed by hnRNP L [CD44 variable exon v5 together with expression constructs encoding Sam68, hnRNP L or both  [CD44 exon v5  . Despiteexon v5  supplemented with 10% foetal calf serum (FCS) (PAA Laboratories). HEK293, HeLa, NIH3T3, and Saos-2, and cells were cultured in Dulbecco's Modified Eagle Media (D-MEM) with glutamax-1 (PAA Laboratories) with 10% FCS. Plasmid transfections were performed as previously described  using GeS-transferase) goat antisera .The following antibodies were used: normal rabbit IgG sc-2027 and anti-Sam68 sc-333 rabbit antisera (Santa Cruz Biotechnology); anti-hnRNP A1 9H10, anti-hnRNP L 4D11, anti-SC35 SC-35, and anti-β-actin AC15 mouse monoclonal antibodies (Sigma); anti-hnRNP A2/B1 DP3B2 mouse monoclonal antibody (Abcam); anti-hnRNP LL rabbit antisera (Aviva Systems Biology); anti-GST  and stained with Coomassie blue . Excised bands were digested with trypsin, and identified by peptide mass fingerprinting. Mass spectra were obtained using the Ultraflex II MALDI-TOF (matrix-assisted laser desorption/ionization time-of-flight) mass spectrometer  in standard MALDI-TOF mass spectrometry (MS) mode. The peak lists were searched against the non-redundant Swiss-Prot protein sequence database using the Mascot  search eLNCaP cell nuclear extracts were obtained using the CelLytic NuCLEAR Extraction Kit (Sigma) according to manufacturer's instructions. Benzonase nuclease (Sigma) was used at 100 units/ml where indicated. Immunoprecipitations were performed using Dynabeads Protein A (Invitrogen) or anti-FLAG M2 agarose (Sigma) according to manufacturers' instructions. Where indicated, antibodies were cross-linked to Dynabeads Protein A according to manufacturer's instructions. Recovered material was resolved by SDS-PAGE and subjected to Western analysis as previously described .Sucrose gradients were run using HEK293 cell nuclear extract, pre-treated or not with micrococcal nuclease (MNase), and recovered material resolved by SDS-PAGE and subjected to Western analysis as previously described .LNCaP, HeLa, and Saos-2 cells were grown and transfected on glass coverslips , and stained with primary and secondary antibodies as previously described . All imaHEK293 cells were grown in 6-well plates (Asahi Techno Glass), and transfected with DNA as detailed in figure legends. RNA was extracted using TRIzol (Invitrogen) and RT-PCR was performed using the One-Step RT-PCR kit (Qiagen) as previously described . DensitoPR conceived of the study, designed and performed experiments, analysed data and drafted the manuscript. CD designed and performed experiments. CB designed and performed experiments. MH contributed reagents and expertise. KE carried out the mass spectrometry. EC contributed reagents and expertise. AB contributed reagents, expertise, and to the preparation of the manuscript. JS designed experiments, contributed reagents, expertise and to the preparation of the manuscript. CR contributed reagents, expertise and to the preparation of the manuscript. HYL conceived of the study, contributed expertise and to the preparation of the manuscript. DJE conceived of the study, contributed reagents and expertise, and drafted the manuscript. All authors read and approved the final manuscript.Sam68 is efficiently immunoprecipitated by its cognate antisera. LNCaP cell nuclear extracts were subjected to immunoprecipitation (IP) using anti-Sam68 rabbit antisera. IP was carried out in the presence or absence (+/-) of antisera to Sam68 or normal rabbit IgG (negative control), with or without (+/-) Dynabeads Protein A, and either with or without (+/-) prior cross-linking of the antisera to Sam68 to the Dynabeads Protein A. Recovered material was subjected to Western analysis with the antisera to Sam68. Western analysis confirmed that Sam68 protein was efficiently immunoprecipitated by its cognate antisera, and that this was most efficient if the antisera were cross-linked to the Dynabeads Protein A prior to incubation with nuclear extracts (compare lanes 3 and 4). Sam68 protein was not pulled down by Dynabeads Protein A alone (lane 2).Click here for filehnRNPs A1 or A2/B1 do not localise within SNBs. (A to D) Representative indirect immunofluorescence images of HeLa cells captured by confocal laser scanning microscopy using antibodies to hnRNPs A1, A2/B1, hnRNP L, hnRNP LL, and Sam68. hnRNPs A1 (A) and A2/B2 (B) exhibit a diffuse nucleoplasmic distribution but do not co-localise with Sam68 to SNBs (arrowed). (C) Sam68 co-localises with hnRNP L to SNBs (arrowed). (D) hnRNP LL exhibits a different subnuclear localisation to hnRNP L and does not co-localise to SNBs. (Bar = 10 μm).Click here for filehnRNP G represses Sam68-mediated CD44 variable exon v5 inclusion. (A) Minigene pETv5 contains the CD44 variable exon v5 cloned downstream of a constitutively-active RSV promoter. Splicing events are shown by broken lines, and arrows show the location of primers for RT-PCR. (B) HEK293 cells were transfected with the pETv5 minigene (150 ng), and expression vectors for hnRNP G or GFP-Sam68 (500 ng). The gel images are representative of at least three independent experiments, from which densitometric assessment of RT-PCR product was performed to obtain means +/- standard error (shown in bar chart). The lane marked M shows the migration of the 1 Kb plus DNA ladder (Invitrogen). Ectopic expression of hnRNP G protein repressed CD44 variable exon v5 inclusion both in the presence and absence of ectopic Sam68 .Click here for file


In the Netherlands, pregnant women without medical complications can decide where they want to give birth, at home or in a short-stay hospital setting with a midwife. However, a decrease in the home birth rate during the last decennium may have raised the societal costs of giving birth. The objective of this study is to compare the societal costs of home births with those of births in a short-stay hospital setting.This study is a cost analysis based on the findings of a multicenter prospective non-randomised study comparing two groups of nulliparous women with different preferences for where to give birth, at home or in a short-stay hospital setting. Data were collected using cost diaries, questionnaires and birth registration forms. Analysis of the data is divided into a base case analysis and a sensitivity analysis.In the group of home births, the total societal costs associated with giving birth at home were €3,695 (per birth), compared with €3,950 per birth in the group for short-stay hospital births. Statistically significant differences between both groups were found regarding the following cost categories 'Cost of contacts with health care professionals during delivery' -76;-25), p < 0.05), 'cost of maternity care at home' , p < 0.05) and 'cost of hospitalisation mother' , p < 0.05). The highest costs are for hospitalisation . Because there is a relatively high amount of (partly) missing data, a sensitivity analysis was performed, in which all missing data were included in the analysis by means of general mean substitution. In the sensitivity analysis, the total costs associated with home birth are €4,364 per birth, and €4,541 per birth for short-stay hospital births.The total costs associated with pregnancy, delivery, and postpartum care are comparable for home birth and short-stay hospital birth. The most important differences in costs between the home birth group and the short-stay hospital birth group are associated with maternity care assistance, hospitalisation, and travelling costs. In comparison with other European countries, the organisation of the Dutch obstetric system is unique, with a high percentage of home births  and a low rate of medical interventions (the rate of Caesarean sections is about 15%) . TraditiThe Dutch obstetric system has received a great deal of attention in the literature . HoweverSeveral studies have examined the economic implications of home births or short-stay hospital births in comparison with a hospital birth -24. HoweThis study sets out to investigate the differences in costs from a societal perspective between low-risk nulliparae preferring to give birth at home and low-risk nulliparae preferring to give birth in a short-stay hospital setting.+0 until 28+6 of pregnancy, week 29+0 until the end of pregnancy, delivery, and the first six weeks of the postnatal period.Cost calculations were performed according to the Dutch manual for costing in health care, a methodological reference for performing costing studies in the Netherlands ,27. ThisThe second step concerns the choice of the cost categories that are measured ,27. In tThe third step of costing is to determine the resources that are used that lead to costs. Contacts with health care professionals, medication, maternity care assistance, medical interventions during delivery, pain control, and hospitalisation were identified as health care costs. Patient and family costs , transportation costs, and extra costs made by responders  were identified as non health care costs. In the fourth stage, the volumes of resources used are measured -28. VoluThe final stage of the cost analysis is calculating the unit costs for each respondent by multiplying the volumes by the unit prices of the resources used -28. The This cost analysis concerns a multi-centered prospective non-randomised controlled study. The individuals participating in the research are grouped according to their preferred place of delivery at home or in a short-stay hospital setting.In the study the following inclusion criteria were applied: the woman is giving birth for the first time (nulliparae), there are no medical indications for secondary care, the woman has the possibility to choose the place of birth , and the woman is fluent in the Dutch language. Recruitment for the study took place on a national level; 100 practices with independent midwives from across the Netherlands were selected at random and participated in recruiting the respondents. The women were informed about the study during their first visit to the midwife (8-10 weeks of pregnancy), and were included in the study if they met the criteria and gave informed consent. Recruitment was carried out from March 2007 to August 2007. Ethical approval was obtained from the Medical Ethical Committee Maastricht (MEC 04-234).The statistical analysis involves analysis of the data collected in the cost analysis. The database was first checked for any erroneous data, by determining the minimum and maximum of the data. The minimum was expected to be zero, which indicated that no contact took place, no medications were used, or no hospital admission was necessary. If the maximum showed extremely high amounts, the database was checked to find out the reason for these outliers. No cases were excluded because of outlying values.The statistical analysis of the collected data is divided into two separate data analyses: a base case analysis and a sensitivity analysis. The data sources collected from the participants showed that not all respondents were complete in registering their data. The reason for performing these two types of analyses was to compare the data as it was received from the participants with the scenario that all participants filled in all sources completely. It is unclear whether women, who did not fully or partially complete the data sources, have more health care consumption.The participants who fully or partly completed the cost diaries and questionnaires were included in the base case analysis. In this analysis some respondents did not complete all items , including data from all participants. To investigate whether the data are distributed normally, histograms were plotted in SPSS, with a normal distribution curve included. It was concluded that the data are not distributed normally, indicating that the data are skewed. Despite the usual skewness in the distribution of costs, arithmetic means are generally considered to be the most appropriate measures to describe cost data ,32. TherIn addition, the mean costs for the actual place of birth were calculated. This statistical analysis was based on three groups .Of the 529 women who gave informed consent to participate in the study, 80 women were excluded because they failed to fill in the first enquiry. Therefore, the study analysed 449 cases. Table Of all women, 31 women (7%) quit participation during follow-up, see Figure Table The number of volumes of resource use per period  are summarised in Table When looking at the different cost categories, the costs for contacts with healthcare professionals are statistically significantly higher in the home birth group  -76;-25), p < 0.05). There are also statistically significant differences between both groups regarding 'costs of maternity care assistance at home' , p < 0.05) and 'costs of hospitalisation mother' , p < 0.05).Furthermore, as is shown in Table The expenses incurred for transportation to the hospital when the delivery started are higher for the women who intended to give birth at home .The results of the sensitivity analysis are shown in Table The total costs of giving birth resulting from the sensitivity analysis are €4,364 for the home birth group, and €4,541 for the short-stay hospital birth group. The total costs for the two groups are higher than in the base case analysis because all 449 respondents are included in the analysis, instead of eliminating those respondents whose data was incomplete. Furthermore, the results in the sensitivity analysis are equal to the results of the base case analysis.The costs of hospitalisation constituted the largest portion of the total costs , as shown in Table Table This is the first article which reports on the first cost analysis into the costs of giving birth in the Netherlands of nulliparous women with different intentions where to give birth: at home or in a short-stay hospital setting. We expected that the costs of home births would be much lower than those of short-stay hospital deliveries. From the results however, it can be concluded that there is no difference in the total costs between the home birth group and the short-stay hospital group. In the home birth group, more costs were spent on maternity care assistance in the postpartum period. This conclusion is in line with the result that the costs of hospitalisation of the mother and child in the postpartum period are higher for the short-stay hospital birth group. In the Dutch obstetric system, women who remain hospitalised after delivery receive fewer days of maternity care assistance at home and therefore receive less reimbursement for maternity care assistance at home. This leads to lower costs for maternity care assistance at home than for the home birth group.Furthermore, the results of the cost analysis have shown that travelling expenses incurred during transportation to the hospital when the delivery started, are higher for women who intended to give birth at home. This may be due to fact these women did not plan to travel to the hospital and are often transferred to the hospital in a later phase of the delivery, when there is more urgency. When looking at the frequencies of using transportation other than the car, 1.2% of women from the home birth group makes use of a taxi and 4.7% is transported to the hospital by ambulance . This indicates that women who intend to give birth at home make use of more expensive transportation more often, leading to higher costs.The results of the cost analysis for the actual place of birth showed a large difference in antenatal costs in "week 29-42" between women who gave birth in secondary care and women who gave birth in primary care. This means that most of the complications during pregnancy arise in the last period of the pregnancy. All respondents were at low risk at the beginning of their pregnancy. When complications occur during pregnancy, their midwife (primary care) has to refer them to much more expensive secondary care. Comparing the results of the analysis of the actual place of birth with the results of the intention-to-treat analysis, a shift can be seen. In the intention-to-treat analysis the costs of a short-stay hospital birth are slightly (but not significantly) higher than the costs of a home birth. In the cost analysis of the actual place of birth, the costs for a short-stay hospital birth are slightly lower than the costs of a home birth. This indicates that the referral rate to secondary care is much higher in the short-stay hospital birth group than in the home birth group, because the expensive care by secondary caregivers will increase the total costs in the intention-to-treat analysis. Further research will be necessary to investigate the difference in referral rates in a short-stay hospital birth and a home birth.Women who opt for a home birth or a short-stay hospital birth have a lower chance for an operative delivery  than women who choose for a hospital birth , while DThe collaboration between midwives and obstetricians has to improve to give adequate information to pregnant women about the differences between home births and short-stay hospital births and the chance for a referral to the obstetrician. Women can make optimal decisions about their place of birth what will probably lead to a positive birth experience.The reason for our cost analysis was to provide insight into the costs of giving birth in the Netherlands of nulliparous women with different intentions of where to give birth. A cost analysis from a societal perspective has not been performed in the Netherlands before. The outcomes of other studies that examined the economic implications of home births and short-stay hospital births  in other countries could not be generalised to the Netherlands, because the Dutch obstetric system is different, with a high rate of home births and low rate of medical interventions -22. FurtRatcliffe  concludeAnderson and Anderson  comparedAlthough the studies discussed here provide insight into the proportion of costs of different birth settings, the results are not comparable to our results, because we focused on the difference in costs between intended home births and short-stay hospital births. Our study does not take the costs of planned delivery in the hospital into account.Some limitations in this study need to be considered. An important limitation relates to the method of data collection. This cost analysis concerned a multicenter prospective cohort study. The initial idea was to perform a randomized controlled trial, in which the place of birth was decided for the women by means of randomization. However, this approach appeared not to be feasible, as Dutch pregnant women do not accept randomization for the place of birth, as we have published elsewhere . TherefoAnother limitation concerns the perspective of the study. The aim of this study was to calculate the total costs of giving birth in the Netherlands from a societal perspective. However, the productivity losses of parents are not included in this study, because a difference in productivity losses between the home birth group and the short-stay hospital birth group was not expected. Besides the health care costs, this study also calculated the patient and family costs, which excluded the research from being studied from a health care perspective.The health care costs included the costs of contacts with health care professionals, medication, maternity care assistance, medical interventions during delivery, pain control, and hospitalization. However, the cost of diagnostic tests such as ultrasounds and CTG monitoring were not separately included in this study. Instead, when women received such a diagnostic test, this test was included in the cost analysis as being an extra contact with the concerning health care provider, in which the time of the appointment was taken into consideration. Therefore, the cost of the diagnostic test itself is included in the calculation of unit costs of visits and it is not possible to present the difference in use of diagnostic tests between both groups.The poor completion of the cost diaries that were used to calculate part of the cost volumes is also considered a limitation of our study. Of the diaries returned, a considerable number were answered only in part; not all weeks were completed. It appeared to be very time-consuming to complete the diaries. Therefore, the missing data were supplied by means of the other data, using general mean substitution.In this research the comparison between different birth settings was reduced to an analysis of the costs. The effects of giving birth in the different birth settings were beyond the scope of this analysis. The consequences of giving birth at home, in a short-stay hospital setting, or in the hospital have been studied by several researchers in the past, and these studies indicated similar consequences regarding the different birth settings. A Dutch study concluded that planned home births are at least as good as planned hospital births for women delivering their first child without medical complications, indicating that the choice to give birth at home is a safe choice . SimilarA final limitation concerns the comparison of our results with the conclusions of earlier studies. As was already explained, these studies differed in methodology and were mostly focused on a comparison between short-stay hospital births and hospital births. Furthermore, it is hard to compare a Dutch short-stay hospital setting with those countries, where they are called birth centres. The characteristics and basic idea of such birth settings may differ between countries.The objective of this study was to give a view of the Dutch obstetric system from an economical perspective. This study provides insight into the societal costs of the two groups of women giving birth for the first time in the Netherlands with different intentions regarding place of giving birth. Because of the high rate of home births in the Netherlands, the obstetric system is currently a topic of debate. In summary, from the results of this cost analysis, it may be concluded that there is no difference in the total costs between low-risk nulliparae who prefer to give birth at home and low-risk nulliparae who prefer to give birth in a short-stay hospital setting.The authors declare that they have no competing interests.The idea for the study was conceived by JN, JS and MH. MH and MB were involved in collecting data. MB participated in the study and performed the statistical analysis. MH, SE, JS participated in its design and coordination and helped to draft the manuscript. All authors edited and approved the final version of the manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1472-6963/9/211/prepub