Background: Preterm birth is a leading cause of perinatal mortality and long-term morbidity as well as the long-term health consequences and cognitive outcomes. Objective: Present study was conducted to determine prevalence and risk factors associated with preterm birth in Ardabil, Iran.Materials and Methods: A case control study was conducted between Nov 2010 and July 2011 in all three maternal hospitals in Ardabil. All the live newborns during the study period were investigated. Of 6705 live births during the study period 346 births occurred in <37 weeks were taken as a case and 589 term neonates were taken as a control group. Data were obtained through review of prenatal and hospital delivery records. Univariate and multivariate logistic regression analysis were applied to obtain magnitude of association between independent variables and preterm birth.Results: The prevalence rate of preterm birth was 5.1%. History of previous preterm birth , hypertension , Oligohydramnios , spouse abuse , preeclampsia , premature rupture of membrane , bleeding or spotting during pregnancy , Hyperemesis Gravid arum , urinary tract infection in 26-30 weeks , , diastolic blood pressure ≤60 mmg  were determined as significant risk factors for preterm birth.Conclusion: Early detection and treatment of diseases or disorders among pregnant women especially hypertension, Oligohydramnios, preeclampsia, bleeding or spotting, Hyperemesis Gravid arum, urinary tract infection, and low diastolic blood pressure as well as the improving health care quality delivered to pregnant women may reduce preterm prevalence rate. Preterm birth, childbirth occurring at less than 37 completed weeks of gestation, is the direct cause for 24% of neonatal deaths , 2. Rateet al found no significant relationships between preterm birth and maternal factors . However, significant relation was observed between maternal age and premature rupture of membranes (PROM) and PROM was also related to preterm birth significantly  were included in a case control study conducted from November 2010 to July 2011. Gestational age was estimated using either the first day of the last normal menstrual period or first trimester obstetric ultrasound for all subjects. Preterm birth, defined as birth occurred in gestational age of less than 37 complete weeks. Termination of pregnancy before 22 weeks of gestation (miscarriage), stillbirth after 22 weeks of gestation, multiple birth, abnormality in uterine, cervical cerclage, women living in other cities, and the women with missing or incomplete prenatal care records were excluded from the study. Including criteria for cases were defined as singleton live births occurred in less than 37 complete weeks. Control group considered as the singleton live births occurred at 37 or more completed weeks of gestation. Of 6705 live births during the study period 346 births occurred at <37 completed weeks and they were considered as cases. Control group (N=589) was randomly selected from singleton live births occurred at 37 or more completed weeks of gestation. In order to improve the accuracy and compensate of possible sample loss, the number of newborns in the control group was approximately twice the cases. That is, per every birth occurred at <37 completed weeks, if possible the next two newborns with gestational age of 37 or more completed weeks were selected as control. Data were collected retrospectively through review of prenatal and hospital delivery records using a self-designed questionnaire. The questionnaire contained socio-demographic factors and maternal and prenatal care characteristics of current and previous pregnancies. Socio-demographic factors included: age, education level, social class, maternal occupation, parents' consanguinity, and area of residence . Maternal and prenatal care characteristics in current pregnancy covered body mass index, maternal height, maternal weight gain after 15 weeks of gestation, gestational diabetes, positive Hepatitis B surface antibody (HbsAb) antigen, hypertension in pregnancy, low systolic blood pressure (≤100), low diastolic blood pressure (≤60), hyperemesis gravid arum, unwanted pregnancy, pre-pregnancy care, urinary tract infection during pregnancy in 6-10 weeks, urinary tract infection during pregnancy in 26-30 weeks, bleeding or spotting, spouse abuse, ferrous and multivitamin use, preterm rupture of membrane, hospitalization in pregnancy, dental problems, chronic diseases in pregnancy, location of placenta, amniotic fluid volume, taking medicine during pregnancy, hard working in pregnancy, infertility history and treatment, mode of delivery, congenital abnormalities, previous contraceptive method, hemoglobin level less than 11g/d in 6-10 weeks of pregnancy, hemoglobin level less than 10.5 g/d in 26-30 weeks of pregnancy, smoking, opium and alcohol consumption.Previous pregnancy and reproductive history characteristics included; gravidity, parity, history of abortion, number of previous abortion, interval from previous abortion, birth interval, history of still birth, history of preterm labor, history of neonatal congenital abnormalities, history of term Low Birth Weight infant, and previous caesarian delivery. The term preeclampsia was used for woman with hypertension (i.e. blood pressure ≥140/90 mmHg) and proteinuria (i.e. ≥300 mg in a 24-hour sample or ≥30 mg/dl in a single sample). Hypertension before pregnancy or experience of hypertension in early stages of pregnancy (before 20 weeks) without proteinuria referred to chronic high blood pressure . UrinaryTreatment of infertility referred to the current pregnancy that was the result of infertility treatment methods and maternal chronic diseases were referred to cardiovascular, hepatic, renal, and lung diseases . HypotenStatistical analysisChi-square, univariate and multivariate logistic regression were applied to analyze data using SPSS version 16. Chi-square was employed to compare distribution of variables between the groups, univariate analysis was applied to evaluate associations between independent variables and preterm birth. In order to highlight important risk factors for preterm birth, multivariate regression analysis was conducted and all the variables displaying significant relationships with preterm birth in the univariate analysis were entered into the model . For the logistic regression, results are reported as odds ratios and 95% confidence intervals (CI) along with P values. SPSS version 16 was used to conduct these statistical analyses. Out of 6705 live newborns delivered, 346 (5.1%) were preterm. Among the variables considered as socioeconomic characteristics, only social class of parents was significantly correlated with preterm delivery . The results for statistical analysis of maternal and prenatal care characteristics of the subjects in current pregnancy in relation with preterm birth are summarized in th week of pregnancy, low systolic blood pressure, polyhydramnion, pregnancy through In vitro fertilization (IVF), pregnancy through ovular stimulator drugs consumption and congenital abnormalities  is appreciably lower than the rates reported for African (12.6%), Asian (9.8%), and some European countries (6.7%) as well as the figures reported for Vietnam (i.e. 11.8%) but is higher than rate of 4.4% reported for Italy % is appr. HoweverExperience of previous preterm birth was identified as the most significant risk factor for preterm birth with odds ratio of 12.7. This is in line with the findings of studies where women with previous preterm delivery were at increased risk for their next pregnancy , 20, 21.Persisting or recurrent intrauterine infection during several pregnancies along with the disorders associated with preterm birth  that tend to last from one pregnancy to the next, might explain many repetitive spontaneous and induced preterm births . Other iet al reported the likelihood of preterm birth to be 2.6 times greater among women with chronic hypertension , may be associated with reduced utero-placental perfusion, prematurity, and low birth weight . StudiesThe data for some variables were collected based on the medical records of the subjects and that; there might be a potential risk for quality of recorded data. Other limitation was that, where the mothers did not remember the date of the first day of last menstrual period accurately, we calculated gestational age by obstetric ultrasound taken in first trimester. To insure the accuracy of data the obstetric ultrasound should be performed by the same person and the same machine for all participants; while it was not possible in our study. History of preterm birth, hypertension and pre-eclampsia, preterm rupture of the membrane, oligohydramnios, spotting and bleeding, urinary infection, hyperemesis gravid arum, and low social status and low diastolic blood pressure were identified as the most important risk factors for preterm birth. Identifying pregnant women at the risk of preterm delivery and proving quality healthcare may decrease the rate of preterm birth and its consequences.


To identify characteristic risk factors of preterm birth in Central and Eastern Europe and explore the differences from other developed countries.Data on 33 794 term and 3867 preterm births (<37 wks.) were extracted in a retrospective study between January 1, 2007 and December 31, 2009. The study took place in 6 centers in 5 countries: Czech Republic, Hungary (two centers), Romania, Slovakia, and Ukraine. Data on historical risk factors, pregnancy complications, and special testing were gathered. Preterm birth frequencies and relevant risk factors were analyzed using Statistical Analysis System (SAS) software.All the factors selected for study , except the history of diabetes were predictive of preterm birth across all participating European centers. Preterm birth was at least 2.4 times more likely with smoking (history or current), three times more likely with preeclampsia, 2.9 times more likely with hypertension after adjusting for other covariates. It had inverse relationship with the significant predictor body mass index, with adjusted risk ratio of 0.8 to 1.0 in three sites. Iron use and anemia, though significant predictors of preterm birth, indicated mixed patterns for relative risk ratio.Smoking, preeclampsia, hypertension and body mass index seem to be the foremost risk factors of preterm birth. Implications of these factors could be beneficial for design and implementation of interventions and improve the birth outcome. Preterm birth (PTB: spontaneous and indicated), defined as delivery before 37 weeks of gestation, is the most common cause of neonatal mortality in developed countries . WorldwiChildren born prematurely have a higher incidence of cerebral palsy, sensory deficits, respiratory illnesses, and learning disabilities compared to children born at term. The morbidity associated with preterm birth often extends to later life, resulting in enormous physical, psychological, and economic costs ,7.In developing countries, accurate and complete population data and medical records often do not exist. Therefore, few international studies regarding preterm birth and neonatal deaths have compared social, economic, or ethnic differences, even though several potential risk factors for preterm birth have been identified, including race, physical environment, nutrition, socioeconomic status, and reproductive age ,8. MaterThe contribution of specific risk factors for PTB in Central and Eastern European counties is unknown. Therefore, the Mother and Child Health Research Network (M&CH RN), as one of the networks of the Association for Regional Cooperation in Health, Science and Technology (RECOOP HST), instituted a retrospective review of delivery records of participating hospitals to identify the risk factors of preterm birth. Its specific aims were to identify the risk factors of preterm birth specific for Central and Eastern Europe (CEE).Eight sites from seven member countries of the M&CH RN were included in the exploratory retrospective study. During the study, two of the centers in Croatia and Poland were removed from the study as they failed to meet the study criteria, mainly due to incomplete data review after the data extraction. Clinical data from 37 661 records of singleton deliveries  were extracted for preterm births (<37 wks.) and full-term births (>37 wks.) between 2007 and 2009 in 6 centers of 5 countries: Czech Republic , Hungary , Romania (Carol Davila University of Medicine and Pharmacy), Slovakia , and Ukraine . The center at Pecs contributed data for two years, 2007-2008. Each center acquired the Ethics Committee approval from their respective institutions.http://www.flexiform.eu/), with data extraction and review completed and confirmed in 2010 by at least two members of each center.Data collection sheets covering 94 parameters distributed among three modules were created collaboratively and approved by each center. The three modules relevant to preterm birth were designed to collect information on: (i) historical risk factors, such as socioeconomic factors, smoking, drug use, and diseases (46 parameters); (ii) problems and interventions in the current pregnancy like vaginal bleeding and cerclage (27 parameters); and (iii) special tests and measurement, such as chorionic villi sampling or progesterone use (21 parameters). The data were entered into an online electronic data entry form using a web based data system . Population based statistical data for each country were obtained from Eurostat, the statistical office of the European Union (http://ec.europa.eu/eurostat) or from the national statistical offices: Czech (www.czso.cz/csu/czso/home), Hungarian (www.ksh.hu/docs/hun/eurostat_tablak/index.html), Romanian (www.indexmundi.com/romania/), Slovakian (www.slovak.statistics.sk/), and Ukrainian (www.ukrstat.gov.ua/). Therefore, for the regression model, site  based data and the national or sub-national population based data were used.The countries participating in the study were classified according to the World Bank classification as high-income countries (HIC) – Czech Republic and Slovakia; upper-middle income countries (UMC) – Hungary and Romania; lower-middle income countries (LMC) – Ukraine; and low-income countries (LIC) – none  for both univariate and multivariate analysis to investigate and compare the relationship of most recognized risk factors of pregnancy with preterm birth. The ten most momentous variables based on relevant preterm birth risk literature and data availability were selected: history of smoking, pre-pregnancy diabetes, chronic hypertension, current diabetes, preeclampsia, progesterone use, current smoking, body mass index (BMI), iron use, and anemia during pregnancy. Univariate analysis was used to examine the relationship between preterm birth (binary variable) and each predictor separately. Multivariate stepwise logistic regression was used to examine the prediction of preterm birth by testing selection from all of the predictors into one model. Stepwise logistic regression was used to screen a large number of variables and to fit multiple logistic regression equations simultaneously.Out of total 37 661 births, 33 794 were term and 3867 were preterm births. Mean percentage of preterm birth was 10.27, maternal age for preterm births was 29.30 ± 5.93 years, and mean maternal age for term births was 28.54 ± 5.39 years . The preThe relationship between preterm birth and each predictor was analyzed initially by univariate analysis of the most significant risk factors . For RomP < 0.0001), Slovakia (P < 0.040), and Ukraine (P < 0.001), while data were missing for Hungary (Budapest). History of smoking or current smoking was significant in four centers (P < 0.0001 each respectively) (except Romania and Slovakia) and was higher in the preterm birth group. At one site only, Czech Republic, the history of diabetes and current diabetes were significant predictors (P < 0.020). Only Czech Republic, Slovakia, and Ukraine recorded higher rates of chronic hypertension in the preterm birth group. Preeclampsia was significantly higher in the preterm birth group for 4 centers (P < 0.001 each respectively) (except Czech Republic and Romania). Iron use was also significantly higher in Hungary (Budapest) (P < 0.0001), Slovakia, (P < 0.020), and Ukraine (P > 0.030). In Czech Republic, Hungary (Budapest), and Ukraine, progesterone use was significantly elevated in preterm birth group . Anemia was a significant predictor in Czech Republic (P < 0.003) and Ukraine (P < 0.001) (BMI was a significant predictor of preterm birth in Czech Republic (< 0.001) .From the multivariate model, 9 of the 10 factors chosen for study were predictive of preterm birth across all these European centers. Only history of diabetes was not predictive presumably because of its low prevalence in all sites. The pattern seen for the various countries was similar. Preterm birth was at least 2.4 times more likely with smoking (history or current) after adjusting for other covariates for 4 out of 5 sites that had data. It was at least three times more likely with preeclampsia after adjusting for other key factors at these sites. Similarly, it was at least 2.9 times more likely with hypertension after adjusting for covariates in 3 out of 5 sites. Preterm birth showed inverse relationship with the significant predictor BMI with adjusted risk ratio of 0.8 to 1.0 in 3 sites. Iron use and anemia, though significant predictors of preterm birth, indicated mixed patterns for relative risk ratio .Data outcome of SAS analysis and PTB frequencies for each country were calculated in relation to the national average . PretermEstimating preterm birth rates and associating risk factors  are essential for determining global incidences of PTB and for developing clinical interventions to reduce the risk of premature birth and related long-term cardiovascular, neurological, and metabolic disorders of both mother and child. Significant risk factors for preterm birth in this study were smoking, diabetes, and chronic hypertension. These factors in conjunction with prior preterm birth and preeclampsia have been strongly correlated with preterm birth and high risk of morbidity due to different mechanisms ,6. PriorThe study identified BMI as one of the most significant factors associated with preterm births in CEE. It is important to consider contextual differences in obesity and preterm birth in the region. The interaction between pre-pregnancy BMI and weight gain during pregnancy could not be assessed in this analysis. Our computation did not take into account the duration of pregnancy. Calculating the weekly weight gain rate to adjust for the length of pregnancy was not possible due to insufficient data on the timing of the first prenatal care. BMI is usually an indicator of nutrition status during pregnancy and has been associated with preterm birth in US and some other developed countries . Obese wSmoking rates of each center were compared to the national rates, except in Ukraine. Four out of 5 sites indicated lower rates of smoking than the national data. This finding may suggest that insufficient attention in the network is being directed toward collecting information on this very important potential behavioral issue. A point to remember is that women tend to hide smoking during pregnancy. Alternatively, women in the network may have been advised not to smoke during pregnancy. There are also the latent factors to help or hinder smoking cessation, enhancing the tendency to conceal it during pregnancy. Smoking was also a strong predictor of preterm birth in our study, perhaps by instigating systemic inflammation and vasoconstriction by nicotine, carbon monoxide, and other toxins in cigarette smoke .A retrospective study by Rebarber et al reported an adverse effect of progesterone use . This st2+) and the oxidized ferric (Fe3+) forms, only catalytic amounts can cause disruption of the cellular macromolecules and redox equilibrium. Importantly, it has been reported (Iron supplementation during pregnancy was a significant risk factor for preterm birth in Hungary (Budapest), Slovakia, and Ukraine. In addition, the use of iron supplements was higher in women with preterm birth than with term birth at all sites except the Czech Republic. It is worth mentioning that there are no clear recommendations across the region regarding iron supplementation during pregnancy. Variation in dosage is anywhere from 60-240 mg/d prescribed from beginning of pregnancy or at the time of first clinic visit, which might be any time during pregnancy. Also, there was no uniformity in using Ferrous or Ferric iron in the centers. As iron readily shuttles between the reduced ferrous , the highest rate in Slovakia (12.1%), and insufficient data from Ukraine, suggesting that different risk factors influence the rates of preterm birth in different countries in the region ,23.Preterm birth has been recognized as a worldwide problem, particularly in LMC where its rate is around 25% compared to 5% in HIC. Moreover, in developing countries, 15% of all neonatal deaths are caused by prematurity and its complications ,25. It hThe relative neglect of preterm birth is linked to data gaps in UMC and LMC and even HIC in Central and Eastern Europe, and consequential prematurity, neonatal morbidity, and mortality. The new estimates shown in this report make a substantial contribution to addressing this deficiency. Widely held perceptions that effective care of the preterm baby requires costly interventions well beyond the health budgets of most LMCs, coupled with concerns that greater attention to preterm birth will draw needed funding away from other devastating maternal and perinatal health problems, have also contributed to the reluctance of policy makers to focus on the problem of preterm birth as global priority ,23,24,27http://www.marchofdimes.org/), International PREterm BIrth Collaborative (PREBIC www.prebic.net) of the World Health Organization, and Department of Reproductive Health and Research , efforts to harmonize the data collection are equally vital to the national, regional, and global programs. Excellent recommendations to improve data have been suggested by Lawn et al (There are many reasons for the inadequacies of preterm birth-related epidemiology in CEE, including patient management by primary health care services, incomplete health-related statistics, and health information systems, lack of preterm birth surveillance registries or imperfect coordination among existing registries and reliance on hospital based rather than population-based studies ,28,29. Ewn et al This study is the first attempt to investigate term and preterm birth using a large collaborative approach in the region and it is part of the learning process with shortcomings of retrospective data collection. One of the limitations was the type of participating sites. Since RECOOP HST Association members are Universities and Academic Research Centers, the study could not provide comprehensive data of the participating CEE countries. We created and used standardized electronic data collection forms to gather information at all sites from paper medical records. Other limitations of the study include the method and limitation of data collection at different sites, and the sampling of one or two sites per country, which does not represent the entire country. We tried to validate the data with available national statistics but the major problem was the imperfect data in the paper based medical records.During the retrospective data collection, it was impossible to gather information on the pre- pregnancy BMI from all sites, but still the study identified BMI as the most significant risk factor in the region. The correlation between obesity and diabetes is well known and the incidence of diabetes in the study is high and comparable to the national rates. However, smoking and high blood pressure are major public health issues in CEE although in the studied cohort the smoking rates and hypertension were lower than the national data. During the retrospective data collection, investigators could not have validated the documented smoking records and limited data were available on blood pressure measurements at the admission and during the hospitalization of the subjects.In conclusion, smoking, BMI, preeclampsia, and hypertension were the most significant risk factors in the region. Comprehensive public health education aimed at behavioral modification and lifestyle change could affect smoking and BMI of the population, which is not only beneficial for the society but it is also likely to have a major effect on the overall rate of preterm birth. Thorough epidemiological data are crucial for any intervention to improve the birth outcomes in the region, so nationwide implementation of electronic birth registration among hospitals is imperative for an effective PTB intervention program.


Some legacy and emerging environmental contaminants are suspected risk factors for intrauterine growth restriction. However, the evidence is equivocal, in part due to difficulties in disentangling the effects of mixtures.We assessed associations between multiple correlated biomarkers of environmental exposure and birth weight.p,p´-DDE) were quantifiable in 72‒100% of maternal serum samples. We assessed associations between exposures and term birth weight, adjusting for co-exposures and covariates, including prepregnancy body mass index. To identify independent associations, we applied the elastic net penalty to linear regression models.We evaluated a cohort of 1,250 term (≥ 37 weeks gestation) singleton infants, born to 513 mothers from Greenland, 180 from Poland, and 557 from Ukraine, who were recruited during antenatal care visits in 2002‒2004. Secondary metabolites of diethylhexyl and diisononyl phthalates , eight perfluoroalkyl acids, and organochlorines , –43 g , and –135 g , respectively; and MOiNP was associated with higher birth weight .Two phthalate metabolites , perfluorooctanoic acid (PFOA), and This study suggests that several of the environmental contaminants, belonging to three chemical classes, may be independently associated with impaired fetal growth. These results warrant follow-up in other cohorts.http://dx.doi.org/10.1289/ehp.1408933Lenters V, Portengen L, Rignell-Hydbom A, Jönsson BA, Lindh CH, Piersma AH, Toft G, Bonde JP, Heederik D, Rylander L, Vermeulen R. 2016. Prenatal phthalate, perfluoroalkyl acid, and organochlorine exposures and term birth weight in three birth cohorts: multi-pollutant models based on elastic net regression. Environ Health Perspect 124:365–372;  A recent pooled analysis of 14 European study populations showed that prenatal exposure to PCB-153 was associated with a significant decrease in birth weight, and p,p´-DDE with a nonsignificant decrease  and dichlorodiphenyltrichloroethane (DDT), or its main metabolite dichlorodiphenyldichloroethylene , eight PFASs, PCB-153, and Study populations. Pregnant women were enrolled between June 2002 and May 2004 during routine antenatal care visits at a) local hospitals or clinics in 19 municipalities and settlements throughout Greenland, b) three hospitals and eight antenatal clinics in Kharkiv, Ukraine, and c) a large central hospital in Warsaw, Poland. Eligible participants were at least 18 years old and born in the country of study. At enrollment, women participated in an interview on time to pregnancy and were invited to donate a nonfasting venous blood sample. The study was approved by the local ethics committees, and written informed consent was obtained from all participating women. Details on the INUENDO cohort study design and data collection have been published  with missing data on maternal age, body mass index (BMI), parity, gestational duration, or infant sex were further excluded.Outcome and covariate ascertainment. Birth outcome data were extracted from hospital maternity records by medical personnel.n = 1,250) (n = 48).Gestational age was assigned based on self-reported date of last menstrual period. The outcome of interest, birth weight, was analyzed for only infants delivered at term  . We did Information on sociodemographic and lifestyle factors and reproductive history was ascertained during the baseline interview. Serum cotinine levels were used as an indicator of exposure to tobacco smoke during pregnancy. Prepregnancy BMI was calculated using self-reported weight and height.Exposure assessment. PFASs and phthalates, along with cotinine and vitamin D, were simultaneously determined in 100-μL aliquots of serum by liquid chromatography–tandem mass spectrometry, following an optimized protocol based on p,p´-DDE were analyzed by gas chromatography–mass spectrometry, as previously described : generalized linear models fit with a hybrid of the lasso and ridge penalty functions (unctions . Ridge pβEN) are estimated by minimizing the residual sum of squares, subject to a weighted sum of the lasso and ridge penalties:A set of elastic net coefficients  and ridge (α = 0) penalties, and λ represents the penalty (or regularization) parameter, with the degree of shrinkage increasing as λ increases for a given α value.where 0 ≤ α ≤ 1 and λ ≥ 0; and p-values of We used cross-validation (CV) to determine the optimal degree of penalization. We tested models over a grid of α and λ sequences, and selected the combination which yielded the minimum mean-squared error (MSE) of prediction from repeated 10-fold CV. CV was repeated 100 times, each time with different data partitions, to achieve more stable selection than a single CV. CV has been shown in simulations to have good sensitivity (true positive rate) of selection, yet can suffer from inflated false discovery rates . TherefoFor the subset of exposures selected via elastic net, and inferred to be important—giving consideration to selection consistency across adjustment models, the covariance test, and the magnitude of the coefficients—we refit multiple-exposure ordinary least squares (OLS) regression models to obtain unpenalized, mutually adjusted coefficient estimates.Before modeling, exposure variables were natural log (ln)–transformed to reduce the influence of outliers. Data below the LOD (0‒27%) were singly imputed from a log-normal probability distribution, dependent on the study population and observed values for other contaminants, and allowing residual variances to vary by population . For regWe identified the minimal sufficient adjustment set using a directed acyclic graph : study population (an indicator variable for participant location), maternal age , prepregnancy BMI (kilograms per meter squared), and parity (≥ 1 vs. 0). Gestational age at birth (weeks) was added in a separate model because it may represent an intermediate variable (or collider)—though it presumably has no direct effect on measured maternal serum levels, it affects the unmeasured cumulative exposure of the fetus, and may also be causally affected by exposures—and there is no consensus on how to adjust for this variable in the literature. In further adjusted models, additional potential confounders were added, possibly leading to over- or unnecessary adjustment: infant sex , maternal height (centimeters), alcohol consumption around the period of conception (≥ 7 vs. < 7 drinks/week), maternal serum cotinine (nanograms per milliliter), and maternal serum vitamin D (nanograms per milliliter). Because of the large proportion of missing data for maternal education, seafood consumption, and season of blood sampling, these variables were included only in a sensitivity analysis.2), and smoking  in straovariate . Lipids ovariate , and fit package .Participant characteristics varied somewhat across study populations . Women fp,p´-DDE . Spearman correlation coefficients between exposures ranged from –0.34 to 0.78 . Exposure distributions differed across study populations for nearly all contaminants . For instance, maternal age exhibited an inverted U-shaped association and parity a positive association with term birth weight, whereas maternal age was positively associated and parity negatively associated  with persistent contaminant concentrations.EN > |1.0|) in the multiple-exposure elastic net regression of term birth weight  (EN > |1.0|) across adjusted models , yielding potentially less reliable multiple-exposure βOLS estimates for differences in term birth weight (grams) when added to the four-exposure model in turn: per 2-SD increase in ln-transformed PFHxS, –19.68 ; PFNA, –17.34 ; PFDoDA, –56.23 .In the primary variable selection analysis, with minimal sufficient adjustment, eight exposures were selected  . PFHxS, d models . These eEN and βOLS) generally decreased slightly upon adjustment for gestational age, and more so upon adjustment for further covariates, with the notable exception of PFOA, which exhibited a more pronounced reduction in further adjusted models  and ΣDiNPom was not selected.The magnitude of associations (βd models . The estq < 0.05) with term birth weight in the primary adjusted model, of which four were also selected (βEN > |1.0|) in the elastic net model. In the models additionally adjusted for gestational age, four were selected in the OLS and five in the elastic net, and three in both models. In the further adjusted OLS and elastic net models, three of the same exposures were selected , and an additional exposure (MOiNP) was selected in the elastic net model .In the comparative analysis with single-exposure OLS regression models, nine exposures were significantly associated , of effect measure modification by study population for MOiNP and PFOA, because the direction of effect estimates was heterogeneous. There was no evidence that infant sex modified associations. Greater reductions in birth weight were observed for overweight/obese compared with normal-prepregnancy-weight women for p,p´-DDE  and for smokers compared with nonsmokers for MEHHP and PFOA  .There was some evidence, although nonsignificant . Only minor differences in coefficients were observed when wet weight rather than lipid-standardized organochlorines were modeled , which appeared to be an artifact of differing exposure ranges and a slight positive slope for Poland and highly negative slope for Greenland , PFOA, and p,p´-DDE) into multiple-exposure unpenalized regression models to obtain estimates of the linear associations between exposures and term birth weight which were not shrunken .To overcome the challenge of assessing associations for correlated exposures, we used a multi-pollutant penalized regression approach to select a subset of the most predictive environmental exposure variables. Elastic net regression forces coefficients of the least predictive variables to zero. We inputted the exposures selected using elastic net regression . Using a more conventional backward–forward stepwise regression yielded generally similar selection results as the elastic net models (data not shown). However, selection accuracy would theoretically be expected to suffer given that exposures are not mutually adjusted at each step. Elastic net would be expected to have even more power in a data scenario with a higher number of exposures and might offer a complementary selection strategy to the single-exposure environment-wide association study (EWAS) approach.Selection based on single-exposure, unpenalized OLS models somewhat coincided with elastic net selection. However, in these models, each exposure was not mutually adjusted for other exposures, and some associations likely reflect false positives that are “tracking” with correlated exposure(s) truly associated with the outcome. For instance, the magnitude of associations for PFOS and PCB-153 decreased markedly when modeled with other exposures (data not shown). We decided to use a penalized regression approach because we expected, based on the correlation matrix, that an OLS regression model including all exposures would suffer from multicollinearity. p,p´-DDE and birth weight . Improving the statistical inference of elastic net models, including explicitly addressing multiple testing, warrants further research and validation.There are caveats to applying a prediction method to an etiologic research question. With two highly correlated variables, elastic net will select a slightly more predictive variable, and shrink the coefficient of the other variable, which may be etiologically relevant. Nevertheless, with conventional  regression modeling, the strength of an association is usually also interpreted as having etiologic implications. Selection based on minimizing prediction error in CV, as performed in the present study, has been shown to overselect predictors. Thus we also applied the recently proposed covariance test for postselection inference , which hh weight . Other sh weight , which an = 4,266) reduced the effect estimates for cord blood PCB-153 and birth weight by 48% and 34%, respectively . We did not have data on gestational weight gain, a proxy of lipid gain, which has been shown in simulations to confound associations between organochlorines and birth weight  but has ectively . For PFAectively .Further, we did not have data on interpregnancy interval, gestational diabetes, preeclampsia, or detailed information on fish intake, possibly contributing to residual confounding. We adjusted for study population to limit bias due to unmeasured confounding, but a corollary is reduced contrast in exposure. We also cannot exclude the possibility that unmeasured (contaminant) exposures confound the observed associations. Furthermore, differential gestational timing of sampling across study populations may have led to less precise and biased coefficients.ex vivo conversion during analysis and storage  which tested 11 and 10 urinary phthalate (including DEHP) metabolites, respectively, found no associations for phthalates and some associations for phenols  of environmental PFAS exposure and birth weight, one reported significant inverse associations for all PFASs evaluated  , observed a mean PFOS plasma level of 35.3 ng/mL and PFOA level of 5.6 ng/mL, higher than levels in the present study; 12.9 and 1.7 ng/mL serum, respectively. A significant association was reported for PFOA  but not PFOS—in line with our findings—and the association for PFOA increased slightly in magnitude upon adjustment for PFOS , PFOA  was not associated with term birth weight, whereas PFOS  was inversely associated, albeit not consistently statistically significant across analyses, and coefficients remained stable in a two-pollutant model (Of the larger studies (d PFHxS) ; anotherz-scores ; and twoz-scores  or PFOA,z-scores . A largefor PFOS . Comparifor PFOS . Inconsint model .n ~ 9,000), including the study populations of this study, although risk estimates displayed heterogeneity across cohorts. Cord serum PCB-153  was significantly associated with reduced birth weight, but p,p´-DDE was not significantly associated: –194 g  per 1 μg/L PCB-153 and –9 g  per 1 μg/L p,p´-DDE. Associations were reported not to change in a two-pollutant model  increased oxidative stress (b) modulation of sex hormone or thyroid hormone homeostasis, which both play a role in development and growth velocities. Cord blood organochlorine compounds and maternal serum PFASs have been found to be inversely associated with thyroid hormone levels (Postulated mechanisms include e stress  and b) me levels . There ie levels . Mid-pree levels . In addie levels .p < 0.05); however, the evidence for potential effect modifiers is generally limited or inconsistent for contaminants assessed in the present analysis.Our results suggest that smoking during pregnancy and prepregnancy overweight/obesity may confer greater susceptibility to the effects of some contaminants on birth weight, although mechanisms underlying these potential interactions remain unclear. We found indications that several environmental contaminant exposures, representing three chemical classes , are independently associated with reduced birth weight, with possible implications for health trajectories. Cautious interpretation is warranted in light of possible confounding due to pregnancy-related pharmacokinetic issues and unmeasured contaminant exposures. We used penalized elastic net regression to assess a mixture of environmental contaminants; this modeling approach may prove useful for similar environmental epidemiology analyses of multiple (correlated) exposures.(601 KB) PDFClick here for additional data file.


Evidence shows that family planning contributes to the decline in child mortality by decreasing the proportions of births that are considered high risk. The main objective of the present analysis was to examine the trends in use of modern contraceptives and their relationship with total fertility rate (TFR) and distribution of births by demographic risk factors as defined by mother's age, birth interval, and birth order at the sub-national level in Ethiopia.Analyses used data from three Demographic and Health Surveys in Ethiopia , which are nationally representative data collected through questionnaire-based interviews from women 15–49 using a stratified, two-stage cluster sampling. First, we examined the trends of and relationship between TFR (in the 3 years before each survey) and modern contraceptive use among currently married women in all administrative regions over the time period 2000–2011 using linear regression analysis. We also examined the relationship between birth risks and under-five mortality using the no-risk group as a reference. Finally, multiple logistic regression analysis was performed to estimate the relationship between the effect of being a resident in one of the regions and having an avoidable birth risk  after adjusting for select covariates including wealth, educational status, residence, religion and exposure to family planning information.Sub-national-level regression analysis showed an inverse relationship between modern contraceptive use among married women and the TFR, with an average decrease of TFR by one child per woman associated with a 13 percentage point increase in modern contraceptive use between 2000 and 2011. A high percentage of births in Ethiopia (62%) fall in one of the risk categories (excluding first births), with wide regional variation from 55% in Gambela to 72% in the Somali region. The multivariate analysis showed women living in the Somali, Afar and Benishangul-Gumuz regions had significantly higher odds of having avoidable birth risk compared to those in Addis Ababa after controlling for observed covariates. The trend analysis showed there was a significant drop in the proportion of births from women above 34 years between 2000 and 2011. There was no significant decline in births to women less than 18 years between 2000 and 2011.A majority of births in Ethiopia fall in one of the risk categories, with substantial region-to-region variation in the percentage of births with avoidable risk factors, Somali and Afar having the highest burden. The analysis indicated that births in the three regions had significantly higher odds of having one of the avoidable risk factors compared to Addis Ababa, and we suggest family planning programmes need to identify differentials of modern contraceptive use at the sub-national level in order to better address coverage and equity issues. Family planning contributes to the decline in child mortality by decreasing the proportions of births that are considered high risk. A majority of births in Ethiopia (62%) fall in one of the risk categories, with substantial region-to-region variation. Births in Somali, Afar, and Benishangul-Gumuz regions had significantly higher odds of having one of the avoidable risk factors compared to Addis Ababa. Family planning programmes need to identify differentials of modern contraceptive use at the sub-national level in order to better address coverage and equity issues.Family planning programmes contribute to improving child survival by reducing the proportion of ‘high-risk’ births –5. GenerEvidence shows that there is a strong negative correlation between levels of contraceptive use and levels of high-risk births and infant mortality. High-risk births account for large proportions of births in developing regions. For example, narrowly spaced births and births to the youngest and oldest mothers account for up to 25 and 36% of births, respectively, in many sub-Saharan countries. Similarly, births to adolescents account for at least 15% of births in sub-Saharan African and Latin American countries  of nearly 2 percentage points since 2000, the country has registered significant progress . Modern However, with an unmet need of 25% and demand satisfied only for 53% of women in 2011, there is substantial unfulfilled demand for modern contraceptives. In addition, the data show that progress is unevenly distributed. For example, the mCPR in the Somali region  in 2011 was 3.8%, a figure well below the national average 11 years earlier. The capital city, Addis Ababa, has the highest mCPR among married women at 56%, followed by Gambela and Amhara regions at 33% each. Unmet need is highest in Oromia at 30% and lowest in Addis Ababa at 11%. With unmet need of 16% and met need of 10%, Afar has the lowest demand at 26%, compared to Addis Ababa, which has a demand of 73% , which is the average number of children that would be born per woman if all women lived to the end of their childbearing years and gave birth to children according to a given fertility rate at each age . The TFRkebele) of the country  the trends in the use of modern contraceptives and their relationship with TFR, 2) distribution of birth risk, 3) trends in birth risk categories across regions, and 4) the relationship between the distribution of birth risk and under-five mortality in Ethiopia.By showing the relative contribution of contextual factors (mainly residence in one of the regions) in reducing avoidable birth risks, the analysis will generate crucial information that will be useful to advocate for prioritisation of family planning programmes, particularly in regions with performance below the national average.The analysis uses data from three Demographic and Health Surveys (DHS) in Ethiopia , which are repeat cross-sectional surveys by design. DHS collects questionnaire interview-based nationally representative data from women 15–49 using a stratified, two-stage cluster sampling that allows for specific indicators, such as modern contraceptive use, to be collected. By design, the survey allows the estimation of regional-level indicators. Questions asked include, among others, background characteristics such as age and education, as well as information about key indicators of maternal and child health including birth history, childhood mortality, and use of family planning methods. The surveys are carried out under the aegis of the Ministry of Health and are implemented by the Central Statistical Agency (CSA). ICF International provides technical assistance as well as funding to the project through the MEASURE DHS project, a project funded by the US Agency for International Development providing support and technical assistance in the implementation of population and health surveys in countries worldwide. The data are freely available from the CSA or the MEASURE DHS project , 11, 12.First, we examined the trends of and relationship between TFR (in the 3 years before the survey) and mCPR (including method mix) in the 11 administrative regions over the time period 2000–2011 using linear regression analysis. We also examined the relationship between birth risks and under-five mortality by calculating the risk ratio with 95% confidence interval, using the ‘no-risk’ group  as a reference.Avoidable birth risk refers to categories numbered 3 to 11 below .Not in any risk categoryFirst birth to mother 18–34 years ageMother's age <18 yearsMother's age >34 yearsBirth interval <24 monthsBirth order >3Mother's age <18 years and birth interval <24 monthsMother's age >34 years and birth interval <24 monthsMother's age >34 years and birth order >3Birth interval <24 months and birth order >3Mother's age >34 years, birth interval <24 months and birth order >3The main outcome of interest in this study was the percentage of births with avoidable birth risk for births in the 5 years preceding each survey. Birth risk categories were defined by three main variables: mother's age at birth, grouped into less than 18 years, 18–34 years, and 35–49 years; preceding birth interval, which is the length in months between previous birth and the index child, grouped into less than 18 months, 18–23 months, and 24 or more months; and birth order, grouped into first births, second or third births, and fourth or greater births. As described by Ross and Stover .Independent variables include region , wealth quintile, educational status  residence , and discussion about family planning in health facilities .Modern contraceptive use in this study refers to current use of modern family planning methods among currently married (or in union) women, which include pills, intrauterine devices, injectables, condoms, female sterilisation, vasectomy, implants, and the lactational amenorrhoea method.Finally, multiple logistic regression analysis was performed to estimate the relationship between the effect of being a resident in one of the regions and having avoidable birth risk, after adjusting for select covariates including wealth, educational status, residence, and exposure to family planning information. Stata 12.1 was used for data management and analysis.Overall, the Ethiopian DHS in 2000, 2005, and 2011 surveyed 15,367, 14,070, and 16,515 women aged 15–49 in 14,072, 13,721, and 16,702 households, respectively, with close to 70% of the sample coming from rural areas in all rounds. Distribution of the sample by region is shown in As shown in p<0.05). As expected, the region with the highest drop in TFR (Amhara) also showed the highest increase in mCPR in 2000–2011, at 26.4 percentage points, while the Somali and Afar regions had the smallest increases in mCPR, at 1.4 and 1.7 percentage points, respectively, showing no statistically significant change over time. The changes in mCPR in Dire Dawa and Harari regions were not statistically significant, apparently because of small sample size, as detailed in Modern contraceptive use among currently married women significantly increased by 21 percentage points, from 6.3 to 27.3%  is the most commonly used modern contraceptive method in Ethiopia, accounting for 72.6% of the method mix in 2011, followed by implants and pills at 12 and 7.4%, respectively. Trend analysis showed that implant use increased (from 1 to 12%) while pill use decreased (from 31 to 7.4%), as did traditional methods (from 1.7 to 1.3%). However, the pattern is slightly different in the Somali region, where pill use is as high as 20% of the method mix and injectables constituted 47% in 2011. Details are shown in p<0.001] between the levels of modern contraceptive use among married women and TFR across the different regions of Ethiopia between 2000 and 2011. The analysis suggests that the TFR declined on average by one child per woman for every increase in the mCPR of about 13 percentage points. It also shows that the TFR would be around 6.2 if there were no modern contraceptive use.As shown in As shown in As Addis Ababa had the lowest percentage of births with short birth interval at 10%, followed by Tigray (15%) and Amhara (15%). As expected, Addis Ababa and Dire Dawa (both city administrations) also had higher proportions of first births, at 38 and 23%, respectively, compared to the national average of 15%.Supplementary Table 2 shows that the proportion of births to mothers older than 34 years was the only single risk factor that showed a statistically significant drop between 2000 and 2011 at the national level, moving from 9.4%  to 7.6% . Looking at regional variations, there was a significant decline in the percentage of births with less than a 24-month birth interval in Addis Ababa, from 17.2%  to 9.9% , while the percentage of first births increased significantly from 26.5%  to 37.9 . It is important to note that the Gambela and Amhara regions had a significantly higher percentage of births to mothers younger than 18 years, at 15.6%  and 15.2% , respectively, compared to the national average of 11%  in 2011.As shown in Our analysis did not show a significant decline in the distribution of avoidable birth risk distribution within regions, except in Addis Ababa, where it dropped from 70.9%  in 2000 to 61.1%  in 2011.As shown in As shown in Over the 11-year study period, a majority of births in Ethiopia fell into one of the risk categories, with substantial region-to-region variation, the Somali and Afar regions having the highest burden. The analysis indicated that births in the Somali, Afar, and Benishangul-Gumuz regions had significantly higher odds of having one of the avoidable risk factors compared to Addis Ababa. This is consistent with previous studies showing low contraceptive use in Somali and Afar regions –15.The present analysis suggests that the TFR declined on average by one child per woman for every increase in the mCPR of about 13 percentage points, which is lower compared to developing countries as a whole (average decrease of TFR by one child per woman for every increase in the mCPR of 17 percentage points) . This diThe majority of births in Ethiopia 62%) fall into one of the high-risk categories (excluding first births), compared to the average for developing countries of 56%  has been characterised by an increase in the use of pills [which have a higher rate of discontinuation ], a decrThe multivariate analysis indicated that women in the Somali, Afar, and Benishangul-Gumuz regions had significantly higher likelihood of having avoidable risk births compared to the reference region (Addis Ababa) regardless of residence, education, religion, wealth, or discussion about family planning in health facilities. Being a rural resident and having no education were significantly associated with higher likelihood of avoidable birth risk, which can be partly explained by the fertility preferences of families of similar background . For exaThere are a few limitations to this study. First, fertility is affected by multiple factors including fertility preference and intention, which may not be affected directly by a mere improvement in access to contraceptives. Second, the cross-sectional nature of the surveys precludes making temporal associations between contraceptive use and birth risk distributions, as both are measured at the same time. Finally, although we have attempted to account for variations in important individual-level factors, we must be cautious in making inferences at the regional level as there is substantial diversity in socio-cultural practices and beliefs within regions.Nevertheless, the consistency of the relationship between the levels of contraceptive use, fertility behaviour of women, and the regional distribution of births with demographic risk factors indicate the possibility of cofounding as an alternative explanation is unlikely.The analyses revealed that there was a significant negative correlation between the level of modern contraceptive use among married women and TFR across the different regions of Ethiopia between 2000 and 2011, suggesting that the TFR declined on average by one child per woman for every increase in the mCPR of about 13 percentage points.A majority of births in Ethiopia (62%) fell into one of the risk categories (excluding first births). There was substantial variation in the percentage of births with avoidable risk factors, however, from 55% in Gambela to 72% in the Somali region and 70% in Afar. The analysis indicated that the Somali region also has a slightly different pattern of method mix in favour of short-acting methods (pills), which could limit impact.The trend analysis showed there was a significant drop in the proportion of births to women older than 34 years between 2000 and 2011 at the national level, while there was a significant decrease in the percentage of births with short birth interval only in Addis Ababa. There was no significant decline in births to women younger than 18 years between 2000 and 2011, so this remains an important risk factor, particularly in regions such as Amhara and Gambela where there are high rates of child marriage.Living in the Somali and Afar regions is significantly associated with having higher odds of avoidable birth risk after controlling for observed covariates including wealth status, residence, education, religion, and discussion about family planning in health facilities.Although Ethiopia has made significant progress in improving modern contraceptive uptake and decreasing high fertility, substantial regional variations persist, pointing the need to refocus family planning programmes to promote more equitable access to a broader choice of effective contraceptive methods throughout the country. Effort should also be made to prevent child marriage, which is often a precursor for teenage pregnancy and high-risk births.Click here for additional data file.


KRAS and glutamic-oxaloacetic transaminase (GOT1/AST1) has been described as part of the metabolic reprogramming in PDAC. The clinical relevance of this novel metabolic KRAS-GOT1 link has not been determined in primary human patient samples. Here we studied the GOT1 expression status as a prognostic biomarker in PDAC. We employed three independent PDAC cohorts with clinicopathological- and follow-up data: a) ICGC, comprising 57 patients with whole-exome sequencing and genome-wide expression profiling; b) ULM, composed of 122 surgically-treated patients with tissue-samples and KRAS status; c) a validation cohort of 140 primary diagnostic biopsy samples. GOT1 expression was assessed by RNA level (ICGC) or immunolabeling . GOT1 expression varied (ICGC) and correlation with the KRAS mutation- and expression status was imperfect . Clinicopathological characteristics did not differ when patients were separated based on GOT1 high vs. low (P = 0.08–1.0); however, overall survival was longer in patients with GOT1-expressing tumors . Multivariate analysis confirmed GOT1 as an independent prognostic marker (P = 0.009). Assessment in univariate (P = 0.002) and multivariate models in the validation cohort (P = 0.019), containing 66% stage IV patients, confirmed the independency of GOT1.Prognostication in pancreatic ductal adenocarcinoma (PDAC) remains a challenge. Recently, a link between mutated We propose the GOT1 expression status as a simple and reliable prognostic biomarker in pancreatic ductal adenocarcinoma. Mortality rates in pancreatic ductal adenocarcinoma (PDAC) have not changed in decades and it remains the fourth leading cause of cancer related deaths , 2, withKRAS mutation in PDAC has been identified [KRAS mutation modulates the associated metabolic pathways by inducing GOT1  [KRAS, mediates a shift in the cancer cell's glutamine-based energy supply system towards other pathways. Currently, the clinical relevance of this novel metabolic KRAS-GOT1 link has not been determined in primary human patient samples.Recently, a novel metabolic function of the highly prevalent entified . Brieflyentified , 22. Thientified  is strikT1 link) , 24. TheKRAS-GOT1 link and performed a biomarker study assessing the clinical and prognostic relevance of GOT1. Our findings identify GOT1 expression as an independent prognostic biomarker in PDAC.Here, we screened a genome-wide dataset of human PDAC for evidence of the KRAS-GOT1 link [KRAS to enzyme mRNA expression levels; however, found no correlation with the two directly linked enzymes GLUD1 and GOT1  [KRAS-GOT1 link has been described at the RNA level [KRAS-genotyped PDAC cohort . Thus, probing for the KRAS-GOT1 link at the RNA and protein level in two cohorts demonstrated that GOT1 is apparently not uniformly expressed; at least not in resected PDAC samples.Since the clinical relevance of the recently described OT1 link  is curreOT1 link  and geneOT1 link , is provt shown) . Due to t shown) , we subsNA level , the ICGP = 0.093; P = 0.049; KRAS-GOT1 ‘linked’ group of patients vs. ‘other’; however, failed to find significant differences  of patients with stage IV disease at time of presentation  to stratify patients into subgroups with differences in outcome. Thus, we consider obtaining the GOT1 status as an intermediate effort prognostic biomarker. Second, obtaining the GOT1 status is possible in primary diagnostic FFPE material, even when stored for years. Specifically, our data demonstrating stability of the antigen well beyond the upper 95% confidence interval of long-term survivors provides evidence that GOT1 assessment is robust. Sharing our immunolabeling protocol that employs an established [Third, for many tissue-based prognostic biomarkers confirmation in a validation cohort is lacking. We validated our findings in an independent cohort. Ideally, such confirmation should be performed in independent laboratories. Fourth, the true challenge for a prognostic biomarker is verification in prospective trials. Such testing will be necessary; however, our retrospective design is not necessarily a limitation but an important starting point for prospective assessment. Yet, even prospective validation cannot overcome the practical hurdle of extrapolating from the population-level estimates to judgments for individual patients [The key dilemma of prognostication in PDAC is that a biomarker needs to stratify patient-subgroups with an overall survival of 3–12 months. Here, we report that assessing the GOT1 status in tumor tissue samples has the potential of becoming such a biomarker. Specifically, GOT1 stratifies patients into subgroups with a median overall survival difference of ~6 months. This striking difference was significant in multivariate models and validated in an independent cohort. At this point, prediction of the disease course in PDAC relies on numerous clinical, radiological and pathological factors. For example, factors such as disease-extent, resectability, and TNM-staging are firmly established ; and numablished  and commpatients , 31. Clepatients , we enviKRAS-GOT1 link reported by Son et al., [KRAS-GOT1 link establishes that of one of the hallmark mutations in PDAC, KRAS, also orchestrates metabolic reprogramming via alteration of glutamate processing. The mutated KRAS mediated cytoplasmic shift via GOT1 hitchhikes downstream enzyme cascades that are critical for redox balance and cell growth [KRAS mutations occur early in the multi-step progress towards invasive carcinoma, the cancer cell may have an initial interest in metabolic reprogramming for survival and mutational maintenance via prevention of catastrophic events; such as those mediated by reactive oxygen species. Thus, sustained growth while at the same time maintaining an optimal redox state is useful for a neoplastic clone and one functional explanation for the KRAS-GOT1 link. Subsequent biological alterations remain currently elusive; clearly some tumors lost their GOT1 expression. The described survival differences based on the GOT1 status are not reflecting the natural course of the disease but are the result of therapeutic interference as well. However, given that neither clinical phenotype, nor proliferation fraction, or other biological features  differed between the GOT1 subgroups, we do not have a plausible biological explanation for the strikingly different survival at this time. Thus, from a basic science perspective further exploration is warranted [The starting point for our study was the recently described  et al., . Brieflyl growth . While tarranted –24 and sarranted , 23.In summary, we report the GOT1 tumor tissue status as an independent prognostic biomarker in pancreatic ductal adenocarcinoma.http://dcc.icgc.org/; last accessioned Sept., 20th, 2013), followed by evaluation in a separate, cohort of PDAC (ULM). These two datasets were combined into a screening cohort. We then assessed GOT1 protein expression in an additional, independent validation cohort.We performed an institutional review-board approved retrospective biomarker study to test the potential impact of GOT1 in PDAC. The study is in accord with the precepts established in the Helsinki Declaration. We explored the PDAC dataset provided by the International Cancer Genome Consortium  whole exome data, b) clinicopathological and follow-up data, and c) genome wide expression data  [cohort is comprised of tumor resection specimens from 122 patients diagnosed between December 1997 and February 2009 [KRAS genotyping was performed using a 5% sensitive pyrosequencing assay on a ProMarkQ24 sequencer  [validation cohort consists of a consecutive series of primary tumor biopsy samples from 140 patients diagnosed between March 2009 and March 2013. The ULM and validation cohort samples consisted of routine diagnostic, formalin-fixed paraffin embedded (FFPE) material stored in a non-air-conditioned room in the basement of our institution. These samples were identified by ICD-O and keyword searches in our laboratory information system and each tumor diagnosis was evaluated by at least two board-certified pathologists. Tumor typing  and grading followed WHO criteria [http://www.cancerstaging.org; last accessioned Dec., 19th, 2013).The ICGC h, 2013) . The preary 2009 . KRAS geGermany) . The valvalidation cohort), we defined GOT1-positive (GOT1+) as cytoplasmic staining in the tumor cells that was stronger than that in the surrounding stroma; the remaining cases were considered GOT1-negative (GOT1–). We employed a commercially available goat-anti-GOT1 antibody  raised against a synthetic peptide corresponding to the C-terminal amino-acids 157–167 of human GOT1  [cohort), we defined GOT1-high when log2-normalized RNA probe-set expression values were above 10.3; the remaining cases were considered ‘GOT1-low’. The cutoff was determined using a stringent non-linear 4th polynomic model (see At the protein level (ULM/ system) . Specifiodel see .validation cohort February 2014. Patients were censored if they were lost to follow-up or if they were alive and well.Medical records were reviewed to extract data on clinicopathologic characteristics and outcome. In the ULM and validation cohort, therapy was administered according to national guidelines . The pri2Χ test for multiple comparisons, or t-test for comparison of continuous variables. The Kaplan-Meier method was used to estimate overall survival. In addition to survival analysis, we modeled probability of death in the screening cohort  and superposed and compared the event curve of the validation cohort (including or excluding stage IV patients). Furthermore, we used uni- as well as multivariate Cox proportional-hazards regression models to analyze survival data. Given survival times, final life status  and one (univariate) or more (multivariate) covariates, the regression models produce a baseline survival curve and covariate coefficient estimates with their standard errors, 95% confidence intervals, and significance levels. The covariates included in these analyses were : sex , age (≥ 65), c/pT (3/4), c/pN (1), c/pM (1), Stage (III/IV), Grade (III/IV), KRAS (mut) and GOT1 . In a second step, we combined factors demonstrating significance in univariate assessments in our multivariate model . Log hazard ratios are provided with the 95% confidence intervals (CI) and statistical significance was defined as P < 0.05. Data analysis was conducted using Prism 6.0  or online available resources .Consisted of Spearman correlation coefficient, Fisher's exact test for dichotomous variables,


The levonorgestrel intrauterine system (LNG IUS) has major advantages and could be a “game-changer” in improving contraceptive choice and use. It faces important challenges, however, including: (1) high commodity cost; (2) often-strong provider resistance to IUDs and difficult programmatic requirements; (3) need for demand creation, including assessing if markedly reduced menstrual bleeding is attractive to clients; and (4) the many requirements for introducing any new contraceptive. A good next step would be a well-focused and multifaceted “learning introduction” to assess the LNG IUS’s potential in several low-income countries, with rapid scale-up if results are promising. The levonorgestrel intrauterine system (LNG IUS) has major advantages and could be a “game-changer” in improving contraceptive choice and use. It faces important challenges, however, including: (1) high commodity cost; (2) often-strong provider resistance to IUDs and difficult programmatic requirements; (3) need for demand creation, including assessing if markedly reduced menstrual bleeding is attractive to clients; and (4) the many requirements for introducing any new contraceptive. A good next step would be a well-focused and multifaceted “learning introduction” to assess the LNG IUS’s potential in several low-income countries, with rapid scale-up if results are promising. Hubacher.See related article by Global Health: Science and Practice, Hubacher makes a good case for the levonorgestrel intrauterine system (LNG IUS) and why donors should purchase it for provision in African family planning programs.,In this issue of The LNG IUS is fueling a rise in IUD use in Europe and the United States.Hubacher cites 3 obstacles to greater donor interest in the LNG IUS: high commodity cost; the belief that currently available method options are sufficient; and doubt about overcoming the barriers that the copper T 380A IUD (Cu T) often faces. He then advances 6 reasons why international donors should buy the LNG IUS: (1) it is on the World Health Organization’s (WHO’s) Essential Medicines List; (2) it is well accepted worldwide; (3) it has a “unique/advantageous delivery system” and important non-contraceptive benefits; (4) it will “activate” some countries; (5) many women in sub-Saharan Africa want to use intrauterine contraception; and (6) more highly effective options are needed.To more fully consider Hubacher’s arguments, we address 5 somewhat overlapping aspects:Costs—absolute commodity cost, commodity cost relative to similar methods (or methods occupying a similar niche), service delivery costs, and opportunity costs—are all, appropriately and necessarily, key considerations for donors, policy makers, and program leaders. Regarding commodity cost, it is relevant to recall that in the early days of implant availability, the public-sector commodity cost (for Norplant) was US$23.80 per set (in 1990 dollars). Today in 2015, thanks to donor volume guarantees to manufacturers, the unit cost of implants has been reduced by two-thirds, to $8.50 per set.It seems unlikely that the LNG IUS would be available to donors below the 2015 unit cost of implants of $8.50 per set.Whether or not—and if so, in what respects—the LNG IUS is an IUD has a number of important dimensions and implications. Hubacher presents a good summary of how the LNG IUS and Cu T IUD differ. But there are also a number of important, programmatically relevant parameters in which the methods are similar. First, both devices require a skilled, unbiased, and motivated service provider in order for clients to receive the method. Second, both require pelvic examination for placement, often a limitation in traditional societies and low-resource settings. Third, both require accurate knowledge by clients—and the Cu T  has long faced a plethora of misunderstandings and myths. Fourth, despite the LNG IUS’s known benefit in reducing abnormally heavy menstrual bleeding and presumed positive effect on anemia, the reduced bleeding or frank amenorrhea it causes could be more problematic in traditional settings where sociocultural beliefs, proscriptions, and/or fears related to menstruation and amenorrhea are more common. None of this means the LNG IUS would not be a very good addition to a country’s method mix; only, rather, that these IUD-related barriers would need to be addressed.,Fostering “unlearning” among medical professionals, even when it is “evidence-based,” is often difficult and time-consuming.Changing providers’ specific understandings and behaviors regarding IUD provision can be particularly difficult. In our experience, providers often have exaggerated but deep-seated concerns related to the triad of sexually transmitted infections (STIs), pelvic inflammatory disease (PID), and infertility, as well as misunderstandings regarding eligibility, believing wrongly that a woman needs to be married or to have had a child to be eligible to receive an IUD.Considerable program effort would be needed to ensure provider biases and client misperceptions about the Cu T IUD do not spill over into LNG IUS introduction efforts.A possible programmatic remedy is suggested in Hubacher et al.’s recent study of the perspectives of 27 IUD service providers from Marie Stopes Kenya.The recent rapid uptake of the LNG IUS in the United States and Europe may not necessarily be predictive of an LNG IUS trajectory in Africa. High-income countries have a more ample supply of service providers, contraceptive methods and services are much more widely and easily accessible to clients, women are more highly empowered and knowledgeable about their contraceptive options, and bias against the IUD is less . Conversely, health workforce and health system constraints in sub-Saharan Africa, particularly for clinical method provision, are unfortunately widespread and problematic, as the recent Ebola outbreak has underscored. This reality is a chief reason that short-acting resupply methods of family planning predominate in the method mix of sub-Saharan African countries.The recent marked increases in implant use in sub-Saharan Africa may also not be predictive of how the LNG IUS might fare, given the ways that provision of these 2 hormonally based methods differs. The implant needs no pelvic exam, does not face provider bias against it, and is more amenable to task shifting to lower cadres, including frontline community extension workers, as is being done in Ethiopia (for insertion). According to the UN’s 2015 multinational tabulations, no sub-Saharan African country has an IUD prevalence above 4%, only Guinea-Bissau and Kenya have levels above 3%, and the majority of countries have levels below 1%.On the other hand, as Hubacher notes, the IUD has never received the extent of donor support and “buzz” in the international family planning community that has been accorded the implant since the Family Planning 2020 (FP2020) initiative and the donor volume guarantee. And yet, even absent that, the IUD’s share of modern method use in 2014–2015 surveys conducted by Performance Monitoring and Accountability 2020 (PMA2020) is 7% in Kenya,Introduction of a new contraceptive method, with plans for its wider adoption and scale-up, is a complex process that requires considerable and sustained work.In our view 5 things will be required for the LNG IUS to become widely accessible and used:A commodity price comparable to that of contraceptive implants.An introduction and ultimate scale-up plan that addresses the many considerations we’ve laid out above (assuming secure supply of needed LNG IUS devices).Identifying, supporting, and otherwise “nurturing” a nucleus of “champions” to become advocates for the method itself and for its introduction at policy, program, system, and service provider levels.Ensuring that the introduction effort is visible to policy makers and opinion leaders in the relevant government ministries, teaching and training institutions, service delivery systems, and professional organizations , as well as to potential clients.Robust outreach and demand creation efforts that include emphasizing the LNG IUS’s non-contraceptive benefits in reducing menstrual bleeding, i.e., it is not “just another IUD.”Achieving scale-up of demand, availability, and access is a longer-term proposition, but one that should be initiated now. Well-conceived, adequately resourced, and well-implemented introduction should ideally start in several countries using import waivers where necessary to acquire the product in places where it is not currently registered. In order to facilitate that start, we describe LNG IUS product options and some of the pathways forward.,http://www.ica-foundation.org/About_the_Programs/Application_for_New_Projects/.)The legacy LNG IUS product, Mirena, currently sells for well upwards of $400 in the United States, which would be prohibitively expensive for public-sector provision in low-income countries. However, an LNG IUS device (with the same amount of hormone and mechanism of action as Mirena) is available free of charge from the ICA Foundation, a partnership between Population Council and Bayer Healthcare Pharmaceuticals.The ICA Foundation provides LNG IUS devices free of charge to governments, NGOs, and multilateral organizations.The ICA Foundation's donations per country have typically been small, largely because the Foundation is unable to provide resources to support other needed aspects of wider introduction, e.g., the costs of shipment, registrations, or waivers, distribution and supply chain management, counseling and skills training, and demand creation. Thus, in the long run, this foundation might not be the best vehicle for a large scale-up of LNG IUS service delivery to tens of thousands (or more) interested clients. However, donations as large as 5,000 to 10,000 units could possibly be provided annually for initial scale-up efforts, concentrating such donations in 3 to 4 low-income countries that have demonstrated interest and with the capacity for larger-scale introduction than has yet occurred.Liletta, a low-cost alternative to the legacy product, is being made available by Medicines360.Although ultimate scale-up of the LNG IUS is a formidable goal, with substantial funding and implementation challenges, we believe the timing is right for smaller, well-focused introduction/demonstration activities to be launched now. These could assess the LNG IUS’s potential among donors, programs, providers, and clients alike. Such initial projects could access the free LNG IUS commodities now available from the ICA Foundation. Alternatively, devices soon to be available via Medicines360 could be used .The time is right to launch small LNG IUS demonstration activities in a limited number of countries to assess the method’s potential in low-income countries.A key part of the introduction strategy would be to support activities in high-quality service delivery situations favorable to IUD provision, such as those using family planning-dedicated providers, mobile outreach service delivery, or social franchising, and with free or very low-cost service provision to clients. On the demand side, having careful evaluation of client acceptability, provider embrace, and method uptake and continuation would be essential in the context of wide contraceptive choice. Mindful of the Cu T’s difficulties, the LNG IUS should be framed as a “new” method, widely used and popular in high-income countries of Europe and North America. Along with high effectiveness and convenience, reduced bleeding and amenorrhea should be specifically promoted as potential advantages, with strong qualitative research to assess that messaging.In Zambia, the dedicated provider approach, which entailed placement of 18 LARC providers in busy, urban-based, public-sector facilities, resulted in provision of 11,000 Cu T IUDs  in 14 months.Price and aggregate commodity cost remain an unresolved sticking point to scale-up, as a 10- to 30-fold cost differential between the LNG IUS and Cu T IUD would understandably represent a large hurdle for policy makers and program funders. Securing a concentrated supply of LNG IUS devices in several countries via the free devices from the ICA Foundation could be a solid, albeit small, start. Hopefully, the imminent availability of Liletta as well as lower cost LNG IUS devices from other manufacturers will foster competition, and commodity cost will accordingly come down substantially. If initial introduction efforts are successful and quantities increase, that could also further decrease cost. And as with contraceptive implants, approaches like a volume guarantee could provide a major boost.Under the conditions described above, the LNG IUS—such a highly beneficial, convenient, and potentially popular contraceptive—could become as widely chosen by women in sub-Saharan African countries as it is now by women in high-income countries of Europe and North America. Moreover, it seems likely that these efforts would also improve use of copper IUDs as well, as was the case in Kenya.In light of the LNG IUS’s considerable game-changing potential, and mindful of these qualifications and amplifications to Hubacher’s call to action, let’s mount a vigorous, well-focused, and multifaceted effort, now.


Background and objectives: Despite growing interest in the role of maternal psychosocial stress as a determinant of preterm birth, no existing work has examined the relation between maternal stress and post-term birth (≥42 weeks). We hypothesize that prolonging gestation past term may represent an adaptive strategy to a suboptimal environment.Methodology: We examined the relationship between exposure to the September 2001 terrorist attacks and odds of post-term birth in California. We calculated the expected odds of post-term birth among conception cohorts of singleton gestations in California between October 1996 and November 2005. We used time series analysis to test for higher than expected odds of post-term birth among the 10 cohorts exposed to the attacks of September 2001 (those conceived from December 2000 to September 2001).Results: The observed odds of post-term delivery among gestations at 33–36 weeks in September 2001 were higher than statistically expected for all race/ethnic and sex groups.Conclusions and implications: Our finding that odds of post-term birth were higher than expected among pregnancies exposed to the September 2001 terrorist attacks in late gestation provides initial support for the hypothesis that exposure to a psychosocial stress during pregnancy may result in prolonged gestation. Post-term ≥42 weeks) birth is associated with increased risk of maternal and neonatal complications , 2, but  weeks biThe exact mechanisms by which maternal psychosocial stress impacts the outcome of pregnancy remain poorly understood. Theory suggests that maternal stress hormones may serve as indicators of environmental conditions, enabling both mother and fetus to adjust adaptively to those conditions , 7. The On the other hand, stress cues later in pregnancy may not trigger parturition because this would result in a viable infant requiring care and nutrition. Birth several weeks prior to term would not conserve maternal resources or promote infant survival, especially in a suboptimal environment. Even late preterm births (33–36 completed weeks) have substantially increased risk of mortality and morbidity compared to term (37–41 completed weeks) births . Furtherlate in gestation (i.e. after the fetus has reached viability), prolonging pregnancy may represent an adaptive response that protects infants until conditions improve or the infant is stronger. In particular, prolonging gestations already at term (37–41 completed weeks) may result in increased post-term birth. It is important to note, however, that risk of both maternal and infant morbidity and mortality increase after ∼41 completed weeks of gestation [This line of reasoning leads us to the corollary that, under suboptimal environmental conditions that induce maternal stress estation , 16. TheWe, therefore, hypothesize that exposure to unexpected environmental stressors late in gestation will be associated with increased risk of post-term delivery. We test this hypothesis by examining the relationship between the terrorist attacks of September 2001 and the odds of post-term delivery among conception cohorts of singleton gestations in California. We chose the terrorist attacks of September 2001 as a stressor because the literature reports that the attacks resulted in severe psychological distress among Americans , 18 and We test this hypothesis separately for male and female gestations among mothers of non-Hispanic white, African American, Hispanic and Asian mothers (i.e. eight groups). We pursued stratified testing because previous research suggests that relations between environmental stressors and birth outcomes differ by sex , 25 and et al. [The California Department of Public Health provided us gestational age , sex, maternal race/ethnicity and date of birth for all singletons yielded by the 110 monthly conception cohorts conceived from October 1996 to November 2005. We excluded births with improbable gestational ages as defined by Alexander et al. . Our stuWe used these data to calculate the monthly odds of post-term birth (i.e. >41 completed weeks of gestation) among the live births yielded by the 110 monthly conception cohorts. We transformed these odds to their natural logarithms to reduce variability in variation over the cohorts and to allow us to express our results in the familiar ‘effect on odds’ metric.in utero to the events of September 2001 (i.e. those conceived from December 2000 to September 2001).Although our argument suggests that delayed parturition will affect stressed pregnancies that have reached term, it remains plausible that pregnancies stressed earlier may, upon reaching term, also benefit from lengthened gestation. We, therefore, test for higher than expected odds of post-term births for all 10 conception cohorts exposed Our argument implies that fetuses in the 37th–41st week of gestation during September 2001 exhibited odds of post-term birth higher than statistically expected under the counterfactual assumption that the terrorist attacks had not occurred. Researchers would typically assume that the statistically expected value under the counterfactual would equal the mean of the odds of post-term birth among a reasonable number of cohorts unexposed to the events of September 2001. This assumption could, however, lead to spurious inferences if a time series of these odds exhibited ‘autocorrelation’ including trends, cycles or the tendency to remain elevated or depressed, or to oscillate, after high or low values. The expected value of such a series is not its mean but rather the value predicted by a model that best expresses the series’ autocorrelation.m predict the odds in month m + 12. Using more, conversely, could have detected and specified weak autocorrelation thereby shrinking the confidence intervals of tests and making it more likely that a small movement away from the counterfactual would appear ‘significant’ [Our test identified and modeled autocorrelation in odds of post-term birth in 110 monthly cohorts conceived from October 1996 through November 2005. This approach derived the counterfactual from autocorrelation in the 10 exposed cohorts as well as from that in the 50 prior and 50 subsequent cohorts. Using fewer cohorts could have provided insufficient power to detect seasonality in which the odds in month ificant’ .We used the strategy attributed to Dickey and Fuller  as well As the first step in our test, we built a Box–Jenkins model of the natural logarithm of the odds of a post-term birth among non-Hispanic white, African American, Hispanic and Asian male and female fetuses in each of the 110 monthly conception cohorts. The general form of the 8 models was as follows.t is the number of infants from the cohort conceived in month t born at gestational age greater than 41 completed weeks. Othert is the number of infants from the cohort conceived in month t born at gestational age less than 42 weeks. d (i.e. odds at month t subtracted from the odds at month t − d). C is a constant. θ is the moving average, or ‘short memory,’ parameter. ϕ is the autoregressive, or ‘long memory,’ parameter. ta is the residual at month t. The ‘backshift operator,’ Bq or Bp, indicates that either θ or ϕ acts on the residual ‘a’ at month t − q or t − p.PostZ scores that estimate their likelihood of occurrence by chance. Our argument that gestations stressed at term (i.e. 37 through 41 completed weeks) will extend to post-term suggests that the Z score for the cohort conceived in December 2000 will exceed 2.6 . The argument that other gestational months will prove ‘sensitive periods’ suggests that a Z score exceeding 2.6 will appear among the months January–September 2001.The residuals of the model have a mean of 0, constant variation and no autocorrelation. We, therefore, express them as Z scores for the residuals from the 10 exposed race/ethnic by sex conception cohorts. The dotted lines show the 99% confidence interval. Results did not support our hypothesis that the cohorts conceived in December 2000 would exhibit scores significantly greater than 0. A ‘sensitive period’ at 33–36 weeks gestation, however, appeared for each of the eight race/ethnicity by sex group. The Z score for every cohort conceived in January 2001 exceeded 2.6. Hispanic males conceived in February 2001 also exhibited a value above 2.6. No residual outside the December 2000–September 2001 ‘window’ exhibited a Z score above 2.6. Z scores below −2.6 appeared for Hispanic females conceived in January and February 2002 and Hispanic males conceived in February 2002.We examined the association between exposure to the terrorist attacks of September 2001 and the odds of post-term delivery among conception cohorts of singleton births in California. Results did not support our hypothesis that cohorts conceived in December 2000 (i.e. at term in September 2001) would exhibit increased odds of post-term delivery. Cohorts conceived in January 2001 (i.e. those in the 33rd through 36th week of gestation when the attacks occurred), however, appeared much more likely to be born after 41 completed weeks.We offer two possible explanations for our findings. The first suggests a ‘sensitive window’ in the 33rd to 36th week of pregnancy in which gestations exposed to stress are ‘programmed’ to be delivered post-term. Most gestations at 33–36 weeks would have been viable even prior to modern medical interventions and might therefore benefit from delayed parturition . However, we know of no such mechanism that would program a gestation stressed at 33–36 weeks to be born to post-term but would not affect gestations stressed at 37–41 weeks.Our second explanation proposes that the real stress felt by Californians was not the events occurring on September 11 itself but, rather, the subsequent fear, uncertainty, and economic collapse that intensified as the implications of the attacks became more clear. Our findings would support the argument that stress experienced at term increases post-term delivery if the stress was felt most strongly in October 2001. The current data does not permit us to discriminate between these two possible explanations for our findings.assessment of stress occurs repeatedly throughout late gestation, enabling the gestation to continue to extend if the stressor is prolonged. A repeated assessment of gestation could help explain our finding that gestations exposed to an initial, acute stressor (September 11) at 33–36 weeks were extended to post-term, especially given the continued stress that the population of California experienced in the weeks following September 11.We should also consider the duration of the population stressor. Our argument presumably applies to acute stressors for which there is reason for the decisional biology of gestation to anticipate improved conditions for delivery of an infant if gestation is prolonged by a period of days or weeks. It remains possible, however, that gestation could be prolonged even in the case of a more chronic stressor in order to deliver a hardier infant, or that the Our findings complement those suggesting that the maternal stress response early in pregnancy terminates the gestations of less fit fetuses to conserve maternal resources for future reproduction at a more optimal time , 8. We pWe recognize that our argument contrasts with that which portrays adverse birth outcomes as a pathological response to stressful environments. We, however, believe that assuming gestation responds adaptively to stress , 7, 12 pWe know of no other study that examines the relationship between maternal exposure to population stressors and post-term delivery. Strengths of our study include the use of a large (i.e. more than 3.6 million), diverse population of births over a period of more than 8 years. An important limitation of this study is that our data comprise monthly conception cohorts, so we are unable to determine exact week of gestation on 11 September 2001. An additional limitation is that measures of gestational age based on LMP can be imprecise , meaningAn alternative explanation for our finding is that medical professionals and/or pregnant women became ‘distracted’ by the events of September 2001, resulting, e.g. in fewer obstetric interventions such as induction of labor or cesarean delivery or a higher likelihood of missed prenatal appointments that may have identified reasons for hastening delivery. This rival explanation would, however, also apply to gestations at term and not, as we found, only to those at late preterm.Findings from this study suggest that maternal exposure to suboptimal environmental conditions during late pregnancy may be linked to prolonged gestation and post-term birth. Our findings support the argument that maternal exposure to stress later in gestation may prolong delivery until conditions improve or until the fetus is strong enough to survive in suboptimal conditions. Future work in this area should examine other types of stressors and potential differences by maternal characteristics, and should try to identify underlying biological mechanisms potentially linking stress to post-term birth.Robert Wood Johnson Health and Society Scholars Program at the University of California, Berkeley and University of California, San Francisco.This work was supported by the Conflict of interest: None declared.