N = 34) repeated the breath holding task in a second session 1 year later. Test-retest reliability of the BHT-task over a 1-year period was high , but none of the executive function tasks was significantly associated with BHT. The rather moderate levels of unpleasantness induced by breath holding in our sample may suggest that other processes  besides distress tolerance influence BHT. Overall, the current findings do not support the assumption of active inhibitory control in the BHT-task in a healthy sample. Our findings suggest that individual differences  should be taken into account when examining the validity of BHT as a measure of distress tolerance.Recent research considers distress (in)tolerance as an essential component in the development of various forms of psychopathology. A behavioral task frequently used to assess distress tolerance is the breath holding task. Although breath holding time (BHT) has been associated with behavioral outcomes related to inhibitory control , the relationship among breath holding and direct measures of executive control has not yet been thoroughly examined. The present study aims to assess (a) the BHT-task's test-retest reliability in a 1-year follow-up and (b) the relationship between a series of executive function tasks and breath holding duration. One hundred and thirteen students completed an initial BHT assessment, 58 of which also completed a series of executive function tasks . A subsample of these students ( Although BHT has not been validated in healthy samples with standard neuropsychological assessment tools, as those applied here, we expected to find positive associations similar to previous studies in which health-related behavior  was predicted. At the same time, we expected breath holding duration to be inversely correlated with anxiety sensitivity as it has been previously reported  completed the BHT session, 58 of which had also completed the entire series of executive function tasks during the computer testing. A subsample of these students  repeated the breath holding task in a second group session  1 year later to examine the test-retest reliability of the BHT-measure.The sample consisted of first year psychology students who participated in a collective psychological testing as a part of their methodology course. Testing consisted of various separate sessions throughout the academic year including a session of group questionnaire completion, a session of group computer testing and one group assessment of BHT. A total of 113 students  as soon as possible every time one of three target letters  appears on screen (“GO” level). At Level 2 (“Go/No-Go level”), participants are asked to respond to two targets (x or y) every time they appear in alternate order, but inhibit their response when the current target is the same as the previous responded target . Level 3 follows the same rule but increases in difficulty since participants have now three targets  to respond (or inhibit their response) to. Levels 2 and 3 are considered to pose increasing inhibitory demands and accuracy at the No-Go trials of these two levels is considered to load on a common inhibitory control factor . At 3-back participants have to respond whenever the current letter is the same as the one presented three trials back. Participants completed two blocks for each level consisting of 30 trials each , ten of which were target trials (to be responded). Total accuracy  was calculated for 0-back and 3-back separately.– Breath holding was assessed in a group manner, requiring modifications to the standard breath holding procedure. Participants were seated in a large auditorium and were instructed to hold their breath after complete expiration . While holding their breath, they were asked to keep their eyes closed. In order to maximize experienced distress the participants were instructed to hold their breath “as long as you can, even if you feel the urge to breathe again.” To calculate BHTs, participants were asked to record the starting time from a timer projected in the room as soon as they held their breath (and just prior to closing their eyes) and the ending time as soon as they breathed in again and opened their eyes. The difference score between the ending time and the starting time was used in analyses.Anxiety sensitivity, i.e., the fear of anxiety-related bodily sensations due to beliefs about possible physical, social and mental consequences, was assessed by the Anxiety Sensitivity Index- 3 , as well as questions regarding participants' experience and compliance with the breath holding task in a retrospective way. Specifically, participants had to rate on a 9-point Likert scale (a) how pleasant/unpleasant they felt during the task , (b) to what degree they followed the instructions precisely , (c) to what degree they tried to hold their breath as much as possible  and (d) whether they could hold their breath even longer . Finally, to control for individual differences in coping behaviors during the task participants had to report their thoughts and possible strategies used during breath holding.Students completed three different sessions of collective testing throughout their first year of studies. The first session was the questionnaire assessment, conducted in a large auditorium where students were seated and completed a series of questionnaires in paper-and-pencil format for about 60 min (those presented here are only a part of the assessment). The second session was the executive function task session conducted in the computer labs of the university. Students were seated in front of desktop computers (in groups of 20), then were given instructions for the three tasks and started completing them at individual pace one after the other. To control for fatigue, the three tasks were counterbalanced, i.e., six orders were created so that each task was presented twice in each position. E-prime 1.0  were calculated to assess test-retest stability. Pearson's r bivariate correlations were conducted to examine the relationships among BHT and psychological variables, while independent sample t-tests were used to investigate possible differences in BHT due to various physiological or demographic factors. Probability level was set to 0.05. All analyses were conducted with STATISTICA 10.0 software (StatSoft/Germany).Pearson's t(32) = 1.07, p = 0.29] and are comparable  with those reported by other studies using the standard BHT procedure with student samples  = 0.67, p < 0.001], see Figure BHTs above 60 s were identified as outliers, resulting in the exclusion of three of 113 participants for the first breath holding test (BHT1) and one of 34 for the retest session (BHT2). Mean BHTs (presented in Table r(110) = 0.06, p = 0.50, or age, r(110) = −0.06, p = 0.50. An independent samples t-test comparing smokers and non-smokers indicated no effect of smoking, t(107) = 0.08, p = 0.93. Among smokers, smoking frequency (daily/weekly/seldom) did not relate to BHT either, rho(15) = −0.18, p = 0.53t-tests, t(110) = 1.69, p = 0.09. T-test analyses indicated no significant sex differences in BHT . Executive functioning tasks did not correlate with BHTs  only. BHT1 did not correlate with body mass index (BMI), r(94) = −0.18, p = 0.075. Even though the correlation is small, it may indicate that anxiety sensitivity influences the relation between BHT and executive functioning. For further exploratory investigation, we split the sample into participants scoring high and low on anxiety sensitivity (median split: 17) and re-examined the correlation between BHT and WCST-number of categories (the index with the highest correlation). For the group with high anxiety sensitivity, a tendency toward a positive correlation was found , whereas no tendency was observed for the low anxiety sensitivity group . The difference between both coefficients was not significant though .None of the questionnaires assessing trait characteristics correlated with BHT. As expected, there was a trend toward a negative association between anxiety sensitivity and BHT1, R2adj = 0.005, F(48) = 1.09, p = 0.36] and none of the predictors significantly explained BHTs .To examine their predictive value in conjunction, WCST-number of categories and total ASI scores were centered and entered as predictors in a multiple regression, as well as their interaction, with BHT as the dependent measure. The overall model was not significant [M = 4.7, SD = 1.5 on a 1–9 scale) did not correlate with BHTs. Neither did the effort indices, nor did the compliance to given instructions correlate with BHT.Regarding participants' experience of and reaction toward the breath holding task, perceived unpleasantness during the breath holding task  than those who did not , t(107) = 2.96, p < 0.005.Twenty-five participants (22.9%) reported to have used a strategy for holding their breath during BHT1. Overall, these strategies fall within four categories: physical actions, relaxation, distraction and mindful focus on breath holding sensations. Physical actions include keeping the nose closed, holding the hands still, closing the eyes and swallowing. Relaxation inducing strategies reported were behaviors like “remaining calm” or “focusing on silence” while the distraction category contains actions to deliberately think of something else or to deviate the attention away from breathing. The category mindful focus on breathing sensations consists of strategies like a direct focus on holding their breath and thoughts about the accompanying sensations. Strategies regarding physical actions and distraction were the ones most used . BHT has been used as a measure of distress tolerance and self-regulation in healthy and clinical samples. However, correlates and determinants of BHT in healthy samples are largely unknown, limiting the construct validity and specificity of this measure. In healthy samples, BHT was interpreted as a measure of distress tolerance  seem to support this suggestion. It could be speculated that a sub-sample of persons scoring high in anxiety sensitivity would show a different result compared to our sample drawn from a general student population in which one can only expect a small proportion of highly anxious individuals.Results of the current study, however, seem to challenge the suggestion of Alpher and Blanton  of an inAccording to our self-report data, experienced unpleasantness during a breath holding task as typically applied in research on distress tolerance and self-regulation is rather moderate in a healthy student sample. It could be hypothesized that executive control resources were not sufficiently activated, as these were not required due to a low level of perceived distress. The standard BHT procedure may be experienced as under full personal control which makes it difficult for strong stress responses to emerge in a highly functioning sample. It therefore seems that breath holding in a healthy student sample may not tap into inhibitory processes but rather be an index of other processes besides distress tolerance, such as physiological and motivational processes. The weak negative correlations between anxiety sensitivity and BHT may provide indications for the effect of such motivational processes although current findings do not allow the examination of such a claim due to the focus on a normal student population with a low prevalence of emotion-regulatory deficits.r = −0.18) of BHT with anxiety sensitivity in this healthy sample is very similar to earlier findings  did not induce a sufficient amount of unpleasantness (distress) in our healthy sample. Thus, motivational and self-regulatory resources were not required to an extent that produces sufficient inter-individual variance to account for differences in BHT. Future research assessing BHT as a measure of distress tolerance should consider increasing the threat value of the task itself by reducing the perceived control of the participant. One such example could be a combination of breath holding along with a mechanically induced full obstruction of breathing (occlusion), which is individually tailored based on previous BHT assessment. In such a paradigm . However, these were not investigated in the present study, nor were these factors typically discussed or investigated in studies on distress tolerance.Firstly, the present study initially assumed a relationship between measures of executive functions and BHT. A wide range of executive function tasks loading on various aspects of inhibitory control  was applied. However, it is important to note that all tasks used neutral stimuli with low emotional significance for the participants, while during BHT the regulation of a rather unpleasant experience is required. The relationship of BHT with the emotional equivalents of these executive function tasks could differ from the one documented in this study, since behavioral control has previously been shown to interfere with simultaneous emotional cues clinical samples. The BHT's stability over a 1-year period suggests that other trait-like characteristics may also be involved. Future research may want to investigate whether other inter-individual differences, e.g., interoceptive sensitivity or other physiological indices of self-regulatory ability, e.g., heart rate variability (Thayer and Friedman, The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Objective. Although several therapeutic strategies have proven to be effective for hydatid cyst of liver, but surgery is still the most common therapy despite its morbidity and mortality. Furthermore, a variety of technique has been recommended for managing the residual cavity after cystectomy. We report here a new technical method for the reconstruction of hydatid cyst residual cavity with using overlapping flaps of liver edges (Vest over Pant). Methods. In this technique after removing the cyst, the edges of one side of cyst cavity were sutured to the base of the cavity using three to four mattress sutures), and edges of other side of liver was overlapped on the dorsal part of previous layer using four to five mattress sutures. Therefore residual cavity dead space was obliterated with two surfaces of cavity. Results. Fifty males were treated by our method. The average cyst volume was 423 ± 110 mL. There was no intraabdominal sepsis, bile leakage, or hepatic necrosis. In follow-up ultrasound study, residual cavities were disappeared one month after operation. Conclusion. Overlapping flaps of liver edges (Vest over Pant) provides easy, safe closure of cyst with preservation of the liver anatomy. The appropriate treatment for hydatid cyst of the liver  is deterThis clinical prospective study was approved by the ethical committee of our university. Participants were all patients consecutively admitted to the general surgery unit of the study hospital for elective surgery of liver hydatid cyst.In approach to liver hydatid cyst, we perform laparotomy with transverse right subcostal incision. Liver mobilization of right hepatic lobe exposing the bare area is performed by dissection of hepatic diaphragmatic attachments. In addition, division of the falciform ligament and peritoneal attachments to the left lateral segment are completed. Operative field is isolated with abdominal packs soaked with 0.5% AgNO3. After that, we proceed to a puncture aspiration of the cyst. Sterilization of cyst content is achieved by infusion of AgNO3 solution for 10 minutes. Each liver cyst is aspirated with a needle and cavity is filled with 0.5% AgNO3 for 10 minutes. Cyst fluid is aspirated again and then opened through a vertical incision to evacuate proligerous membrane. The cyst content is evacuated and cavity evaluated for biliary communications. For intramural cyst, we use cystectomy. Biliary communications are closed with direct suture. Then, the edges of one side of cyst are sutured to the base of the cavity using three to four mattress sutures with chromic 0 designed for liver suturing , and edgFifty males were treated by our method. Ages ranged from 20 to 45 years. The clinical findings were abdominal mass in 20 (40%) cases and abdominal pain in 30 cases (60%). Cyst size was 82.2 ± 10.3 mm (55–145 mm). The described technique has been successfully used in 50 cases. All of the patients had single cyst. In 46 patients, cysts were located in right lobe. The average cyst volume was 423 ± 110 mL (300–650 mL). In two patients, biliary communications were closed by direct sutures. There was no intraabdominal sepsis, bile leakage, or hepatic necrosis. In follow-up ultrasound study, residual cavities were disappeared one month after operation. In one year followup, there was no fluid collection or recurrence of disease in ultrasound study.Hydatid cysts represent the liver stage of the dog tapeworm Echinococcus Granulosus. The disease is endemic in many countries such as Iran. The appropriate treatment of hydatid cyst is determined by several factors and is surgical or percutaneous drainage with intracystically injected scolicidal agents and chemotherapy. However there is considerable controversy about the management of the residual cavity. Many promising surgical techniques have been suggested through the years, but none has proven to be accurate enough, and their clinical relevance has been questioned. Cystectomy is considered the least traumatic method for hydatid cysts excision, but there is typically a larger residual cavity that can lead to a number of life-threatening complications such as suppuration, recurrence, and biliary fistula. Suture obliteration, omentoplasty, introflexion, double breasting, simple closure, deroofing, and tube drainage –11 are t


Introduction. Preterm birth is the leading cause of newborn deaths and the second leading cause of death in children under five years old. Three-quarters of them could be saved with current, cost-effective interventions. The aim of this study was to identify the risk factors of preterm birth in a secondary care hospital in Southern India. Methods. In the case-control study, records of 153 antenatal women with preterm birth were included as cases. Age matched controls were women who had a live birth after 37 weeks of gestational age. Gestational age at delivery and associated risk factors were analyzed. Results. The preterm birth rate was 5.8%. Common risk factors associated with preterm birth were hypertensive disorders of pregnancy (21.4%), height <1.50 m (16.8%), premature rupture of membranes (17.5%), and fetal distress (14.9%). Mean birth weight for preterm babies was 2452 grams while the birth weight for term babies was 2978 grams. Conclusion. The commonest obstetrical risk factor for preterm birth was hypertensive disorders of pregnancy and nonobstetrical risk factor was height <1.50 m. The percentage of preterm birth was low, comparable to developing countries. Preterm birth (PTB) is the leading cause of infant morbidity and mortality in the world. The World Health Organization (WHO) defines preterm birth as any birth before 37 completed weeks of gestation or fewer than 259 days since the first day of woman's last menstrual period (LMP).In developing countries, the main causes of preterm births include infectious diseases and poor availability and accessibility of health care resources. In high-income countries, the increase in the number of preterm births is linked to conception among older women and increased number of multiple pregnancies as a result of usage of fertility drugs. In some developed countries, medically unnecessary inductions and caesarean section deliveries before full term also increase preterm birth rates. In rich and poor countries, many preterm births remain unexplained , 2.Approximately three-fourths of perinatal deaths occur in foetuses that are delivered at <37 weeks, and about 40% of these deaths occur in those delivered at <32 weeks. In addition to its contribution to mortality, preterm birth has lifelong effects on neurodevelopmental functioning such as increased risk of cerebral palsy, impaired learning, and visual disorders and an increased risk of chronic disease in adulthood [Defining risk factors for prediction of preterm birth is a reasonable goal for several reasons. First, identification of at-risk women allows initiation of risk-specific treatment. Second, the risk factors might define a population useful for studying specific interventions. Finally, identification of risk factors might provide important insights into mechanisms leading to preterm birth. Yet, data regarding preterm births and risk factors are not routinely collected in hospitals. Therefore, to obtain insight into the risk factors for PTB, a secondary care hospital was chosen in Southern India for the study.Data of births in Dr. TMA Pai Hospital, Udupi, during the period from January 2010 to May 2013 were used for this study. Approval for the study was taken from the Institutional Ethics Committee. The births at Dr. TMA Pai Hospital are registered in a log book in the labour room. All live births during the period from January 2010 to May 2013 were included into this study. Potential cases were all women who were recorded in the labour log book as giving birth at a gestational age between 28 and 37 weeks. The hospital numbers of these women were taken from the log book and the patient records were collected from the hospital medical records. Data from these records were used to fill the questionnaire.For every case record, two control records were obtained. Controls were all women who had a live birth after 37 weeks of gestational age. The controls were age matched. For every case, two controls two years older or younger than the case were selected by simple random selection. This selection was done by using a random number table.There were in total 238 cases of preterm births during the study period. Out of which 17 babies were still born, 47 records were not found, and 20 cases had inadequate data and hence were excluded from the study so the final analysis was done for 154 cases.The gestational age was assessed by using date of last menstrual period and confirmed by ultrasound in the records. The criteria used for defining the different clinical conditions are shown in The questionnaire used for data entry was divided into six sections: “background information of the mother,” “medical history,” “current pregnancy details,” “baby details,” “details of previous conceptions,” “medical disorders complicating current pregnancy,” and “investigations.” In the “background information of the mother” section, information was obtained about the age, occupation, height, and the body weight at the beginning of the pregnancy. In the “medical history” section, information was obtained about family and medical history of the pregnant woman.Risk factors found in the women were categorized into twelve categories: “fetal distress”; “failed induction”; “hypertensive disorders of pregnancy”; “malpresentation and multiple pregnancy”; “previous uterine scar”; “antepartum haemorrhage (APH)”; “PROM”; “medical disorders”: anemia, hyperthyroidism, hypothyroidism, Rh negative with ICT positive, and Rh negative; “oligo/polyhydramnios”; “gestational diabetes mellitus (GDM)”; “cephalopelvic disproportion”; and “others”: manual removal of placenta, baby hydrocephalus with absent posterior vault, bladder injury, and fetal polycystic kidneys.Univariate logistic regression analysis was used to examine the association between preterm birth and risk factors. The univariate association of risk factors with preterm birth was approximated by determining the odds ratio. The statistical package IBM SPSS Statistics 20.0 was used for all statistical calculations. Odds ratio was calculated for the risk factors associated with preterm and term deliveries. Univariate and multivariate logistic regression analysis were used to examine association between risk factors and preterm and term birth. Statistical significance was set at 5% level. Ninety-five percent confidence intervals for the odds ratio were calculated.Among the 4,137 antenatal admissions during the study period, 238 were admitted with preterm labour. The preterm birth percentage was 5.8%.The baseline characteristics are described in The mothers in both the case and control group were healthy, with very few preexisting medical conditions as shown in Risk factors associated with preterm delivery using univariate analysis are presented in In the control group the most common complication was fetal distress, found in 53 (15.9%) women, followed by failed induction 45 (13.5%), GDM 27 (8.1%), and previous uterine scar 27 (8.1%).Significant associations were found between hypertensive disorders of pregnancy, height <1.50 m, PROM, oligo-/polyhydramnios, threatened abortion, twin gestation, and preterm birth.The present study has shown that preterm delivery was significantly associated with hypertensive disorders of gestation, height <1.50 m, PROM, oligo-/polyhydramnios, threatened abortion, and twin gestation.Hypertensive disorders of pregnancy were present in 21.4%, which were the commonest obstetrical risk factor. The finding was contradictory to the study done by Shresta et al.  in whichAnalysis of nonobstetrical risk factors revealed height <1.50 m as a significant risk factor for preterm birth. Short women are more likely to have a small pelvis, which can lead to obstructed labour. Intrauterine growth restriction is also more likely.In late pregnancy, placenta previa and placental abruption are often associated with vaginal bleeding and often lead to preterm birth (PTB) . In thisMultiple gestations, accounting for only 2-3% of infants, carry a substantial risk of preterm delivery and result in 15–20% of all preterm births. Nearly 60% of twins are born preterm. About 40% of twins will have spontaneous labour or PROM before 37-week gestation, with others having an indicated preterm delivery because of preeclampsia or other maternal or fetal disorders [According to Krupa et al.  vaginal Extremes in the volume of amniotic fluid—oligo- or polyhydramnios—are associated with preterm labour , althougMaternal demographic characteristics associated with preterm birth include low socioeconomic and educational status, low and high maternal ages, and single marital status , 10. ObsBased on data from 184 countries, the global average preterm birth rate in 2010 was 11.1%. Preterm birth rates varied widely between countries. At a national level, the estimated preterm birth rate ranged from about 5% to 18%. The highest rates of preterm birth were in south-eastern Asia and sub-Saharan Africa . Studies revealed incidence to be higher in developing countries than in developed countries . India hThe commonest obstetrical risk factor for preterm birth was hypertensive disorders of pregnancy and nonobstetrical risk factor was height <1.50 m. Significant risk factors for preterm birth found in this study were gestational hypertension, height <1.50 m, preterm rupture of membranes, threatened abortion, and twin gestation.


Background: Tobacco-smoke, airborne, and dietary exposures to polycyclic aromatic hydrocarbons (PAHs) have been associated with reduced prenatal growth. Evidence from biomarker-based studies of low-exposed populations is limited. Bulky DNA adducts in cord blood reflect the prenatal effective dose to several genotoxic agents including PAHs.Objectives: We estimated the association between bulky DNA adduct levels and birth weight in a multicenter study and examined modification of this association by maternal intake of fruits and vegetables during pregnancy.32P-postlabeling technique in white blood cells from 229 mothers and 612 newborns. Maternal diet was examined through questionnaires.Methods: Pregnant women from Denmark, England, Greece, Norway, and Spain were recruited in 2006–2010. Adduct levels were measured by the 8 nucleotides; Spearman rank correlation coefficient = 0.66, p < 0.001). Cord blood adduct levels were negatively associated with birth weight, with an estimated difference in mean birth weight of –129 g  for infants in the highest versus lowest tertile of adducts. The negative association with birth weight was limited to births in Norway, Denmark, and England, the countries with the lowest adduct levels, and was more pronounced in births to mothers with low intake of fruits and vegetables  compared with those with high intake Results: Adduct levels in maternal and cord blood samples were similar and positively correlated . Environ Health Perspect 121:1200–1206;  Measurement of biomarkers in cord blood improves exposure assessment and may improve our understanding of biological mechanisms during this critical window of exposure and vulnerability  . All samples from Greece, Spain, and Norway and the Danish samples collected in 2006–2007 were analyzed in Budapest, Hungary (61% of the samples); the Danish samples from 2009 were analyzed in Stockholm, Sweden (21%); and the samples from England were analyzed in Maastricht, the Netherlands (18%). The interlaboratory comparison study showed a very high repeatability between two of the three laboratories, whereas the adduct levels measured in the third laboratory were consistently 3.7 times lower than the mean levels determined by the two other laboratories. Differences in DNA adduct determinations between laboratories normally occur because of the complicated multistep and sensitive procedures used for the detection of the adducts, and interlaboratory studies are therefore necessary. A correction for the laboratory factor of 3.7 was thus applied to the samples analyzed in Stockholm . A sensitivity analysis that included or subsequently excluded these samples gave similar results, so all analyses were based on the total study population including the corrected data.DNA was isolated centrally from ~ 0.5-mL aliquots of buffy-coat samples using Qiagen Midi Kit catalog no.13343  with some modifications . Levels rotocols . The pro8 unmodified nucleotides (n/108 nt).The individual level of DNA adducts was obtained as the average of at least two independent measurements. The detection limit of the assay was approximately 0.1–0.3 adducts per 10Statistical analysis. We performed linear regression models with a random effect for country to estimate the difference in mean birth weight (grams) associated with bulky DNA adduct levels in cord blood. Furthermore, we estimated associations with head circumference (centimeters) and gestational age (completed weeks) at birth. Low birth weight  and preterm  were too uncommon to estimate associations with bulky DNA adduct levels.n = 205), middle (n = 203), or high (n = 204) .DNA adduct levels measured in cord blood were modeled as categorized according to tertiles: low , infant sex, maternal prepregnancy body mass index , parity , maternal age (years), maternal ethnicity , self-reported maternal active smoking at the end of pregnancy , self-reported maternal exposure to secondhand smoke (SHS) during pregnancy , ethylene oxide–hemoglobin (Hb) adduct levels in cord blood , mode of delivery , dietary supplements , maternal consumption of fruits and vegetables during pregnancy , season of delivery , country of delivery, and maternal education  as a marker of socioeconomic position. In addition, we estimated basic adjusted associations using models that included country, maternal smoking at the end of pregnancy, sex, and gestational age only.We examined the effect of different degrees of adjustment for potential confounders on the association of bulky DNA adduct levels on birth weight. Potential confounders selected Given that bulky DNA adduct levels in cord blood reflect a steady state between DNA damage and repair during the previous few months , we clasn = 54) with a total energy estimate of < 500 or > 6,000 kcal/day were excluded from adjusted analyses and analyses of effect modification by diet  was categorized as high or low according to overall and country-specific median levels as a proxy measure of the consumption of nutrients that might be protective against genotoxic activation processes. We also classified mothers according to their intake of fruits high in vitamin C and other antioxidants . Women ( by diet .n = 367), which had low average levels of adducts, were compared with outcomes in Southern European countries , which had higher average adduct levels.In addition, we performed a meta-analysis by country to derive country-specific effect estimates of associations between cord blood DNA adduct levels and birth weight. Pregnancy outcomes in Northern European countries . We used an alpha level of 5% for statistical significance. All statistical analyses were performed using Stata S.E. version 12.1 .To estimate the association between bulky DNA adduct levels and term birth weight, we repeated the main analysis after excluding preterm deliveries (Study population. The study population was composed of neonates from Denmark (33%), Spain (29%), England (18%), Greece (11%), and Norway (10%). Mothers were predominantly white Europeans, multiparous, and nonsmoking (p = 0.13). The difference in fruit and vegetable intake between Southern and Northern European mothers was also not statically significant after adjustment for individual total energy intake. Differences were smaller for fruits high in vitamin C  and cord blood (n = 612) samples had detectable levels of adducts. Median levels of adducts in paired maternal and cord blood samples were similar . Cord blood adduct levels ranged from 0.6 to 87.5 (adducts/108 nt) and were significantly positively correlated with maternal levels .8 nt; range, 0.8–87.5) than from Northern Europe , although an opposite pattern was observed for birth weight  . Median adduct levels in the children of nonsmokers also differed significantly (p < 0.001) among the individual countries . The same pattern of higher adduct levels in children from Southern Europe than those from Northern Europe was found for children born to nonsmokers without exposure to SHS.Bulky DNA adduct levels were higher in children from Southern Europe  . The dif8 nt; range, 0.6–73.9) was higher than the median level in the 541 children of mothers who never smoked or who quit before the last 4 months of pregnancy . Adduct levels were lowest in the 505 children born to mothers who never smoked during their pregnancy .The median bulky DNA adduct level in the 71 children born to mothers who actively smoked at the end of their pregnancy (10.9 adducts/10Adduct levels in cord blood and birth outcomes. Higher levels of bulky DNA adducts in cord blood were associated with lower birth weight (n = 612), the estimated difference in mean birth weight for infants in the highest tertile versus the lowest tertile of adduct levels was –110 g , based on basic adjusted models that included country, maternal smoking at the end of pregnancy, sex, and gestational age only. The corresponding association was similar when restricted to the 541 mothers who did not smoke during the last 4 months of pregnancy , but was slightly stronger when restricted to the 505 mothers who did not smoke at any time during the pregnancy .h weight . For then = 409), the estimated difference in mean birth weight associated with the highest versus lowest tertile of DNA adduct levels was –129 g  (n = 390), the estimated mean difference was –139 g . When we further adjusted for acrylamide Hb adduct levels in cord blood  the associations remained practically unchanged .After further adjusting for maternal age, prepregnancy BMI, exposure to SHS, maternal education, ethnicity, intake of fruits and vegetables, delivery type, and season of delivery  . After aWhen 26 preterm births (< 37 weeks) were excluded, the estimated difference in birth weight for infants in the highest versus lowest tertile of adduct levels was –139 g .p = 0.08, n = 530) and after further adjustment for maternal age, prepregnancy BMI, exposure to SHS, maternal education, ethnicity, intake of fruits and vegetables, delivery type, and season of delivery .Bulky DNA adduct levels were negatively associated with head circumference based on the basic adjusted model  based on the further adjusted model. When restricted to vaginal deliveries , the estimated difference in gestational age was –0.54 weeks , compared with a difference of 0.15 weeks  for deliveries by cesarean section (n = 166).The estimated difference in mean gestational age at birth for infants in the highest versus lowest tertile of adduct levels was –0.29 weeks  based on the basic adjusted model. However, there was a nonsignificant positive association among the 245 Southern Europeans . Estimates based on adjusted models also indicated negative associations for the Northern European countries, but not for the Southern European countries , and when imputed data were used to include individuals with missing information on covariates (results not shown). Therefore, although birth weight was lowest and adduct levels were highest in Greece and Spain . Consumption of dietary supplements during pregnancy was common in both Northern and Southern Europeans  and did not appear to modify or confound associations between adduct levels and birth weight (data not shown).nificant . The estWe measured levels of bulky DNA adducts in white blood cells from cord blood in a large multicenter European prospective general population study and found that higher adduct levels in cord blood were significantly negatively associated with birth weight. The negative association was observed among newborns from England, Denmark, and Norway, who had the lowest average adduct levels, but was not evident among Southern Europeans, who had the highest mean adduct levels. The negative association with birth weight was stronger among the children of mothers with low versus high intakes of fruits and vegetables.Tobacco smoke contains PAHs and other DNA adduct–forming compounds in addition to other potentially harmful compounds, and active maternal smoking is a recognized risk factor for reduced fetal growth . Furthern = 449 samples, Poland) , bulky DNA adducts      , and str Poland)  support  Poland) , BPDE–DNregnant)  and on B, China) . Resultsin utero possibly related to induction of P450 enzymes, global DNA methylation changes, induction of apoptosis, altered gene expression, cellular mutations, or oxidative stress , possible mechanisms may include binding to aryl hydrocarbon receptor (AhR) and/or other receptors causing endocrine disruption, altered placental growth, decreased placental exchange of nutrients and gases e stress . Develope stress .Our findings suggest that maternal intake of fruits and vegetables may modify the association between bulky DNA adduct levels and birth weight. This finding is consistent with the findings of a questionnaire-based study evaluating dietary BaP .Children from Southern Europe had, on average, higher bulky DNA adduct levels and lower birth weight than Northern European children. However a negative association between adducts and birth weight was found only in the Northern Europeans. The same pattern of higher bulky DNA adduct levels in Southern Europe than in Northern Europe has been found in adults , and thiA key strength of the present study is the measurement of bulky DNA adducts in cord blood, which enabled a more accurate evaluation of the biologically effective dose of genotoxic agents resulting from complex environmental exposures than estimates based solely on external exposure assessment. Large biomarker-based studies are rare because of their costs and complexity. Standardized protocols were developed  and appl32P-Postlabeling is considered to be the most suitable and sensitive method for assessing total DNA damage resulting from exposures to unknown, complex mixtures of genotoxic compounds. We used the validated in vitro BPDE–DNA standard in each 32P-postlabeling session in duplicate for normalization of the DNA adduct levels. Each DNA sample was analyzed at least twice. The interlaboratory comparison study, the use of common protocols, and use of a validated BPDE–DNA standard minimized the potential for measurement error. Thus, we believe that measurement error of the method was properly handled and is not a special concern in our study.in utero formation of bulky DNA adducts also may affect prenatal growth, and that this potential effect may be reduced by high maternal fruit and vegetable consumption.Low birth weight is an important outcome because it is associated with greater risk of neonatal mortality, hypertension and cardiovascular disease, diabetes, certain cancers, reduced and/or delayed postnatal growth, and cognitive development . Our fin


Preterm birth is associated with painful procedures during the neonatal intensive care unit (NICU) stay. Full-term newborns can also experience pain, following surgery. These procedures can have long-lasting consequences. It has been shown that children born preterm show pain responses and cardiac alterations. This study aimed to explore the heart rate reactivity to pain in 107 subjects born either preterm or full-term who were between 7 and 25 years old at testing. We also evaluated the effect of pain experienced at birth, as represented by a longer NICU stay, time under ventilation, and surgery at birth. Participants were asked to immerse their right forearm in 10°C water for 2 minutes. Electrocardiograms were recorded at baseline and during the immersion procedure. Full-term subjects showed a stable increase in heart rate throughout the procedure, whereas preterm ones showed a strong increase at the beginning, which decreased over time. Also, preterm and full-term subjects who experienced pain at birth showed higher resting heart rate, stronger sympathetic activity, and lower cardiac vagal activity. Our study demonstrated a long-term impact of a long NICU stay and surgery at birth on cardiac autonomic activity. This could lead to impaired reactions to pain or stress in later life. In the last decades, advances in knowledge and improvements in treatment options have increased survival rates for preterm infants . HoweverOnly a few studies evaluated heart rate changes in reaction to a painful procedure and have yielded contradictory results , 15. TheOne hundred seven French-speaking subjects aged between 7 and 25 years (mean: 14.6 ± 4.5 years) participated in the study. Healthy control subjects were recruited through local ads. All preterm subjects born at the Centre Hospitalier Universitaire de Sherbrooke (CHUS) or at the Royal Victoria Hospital of Montreal and full-term subjects that underwent surgery at birth at the Montreal Children's Hospital received a letter inviting them to participate in the study. Recruitment was difficult, as only 8% of the subjects contacted by mail sent back a positive answer. All subjects were pain-free at the time of experiment and none of them had taken analgesics or any other medication prior to testing. None of the participants presented any cognitive problems, renal or hepatic failure, cerebral palsy, or epilepsy. The hospitals' review boards approved the study protocol and the experiments took place at the Clinical Research Center of the CHUS, Sherbrooke, Quebec, Canada, and at the Montreal Children's Hospital, Montreal, Quebec, Canada. For subjects under 18 years old who accepted to participate in the study, parents gave their informed consent for their child's participation. Participants 18 years and above signed the consent form before the experiments.Medical birth files were revised after experimental pain testing by two of the authors (Mélanie Morin and Louis Couturier) to determine the extent of painful procedures experienced at birth for preterm subjects and the type of surgery for full-term participants. Since the information on the exact number of painful procedures experienced at birth was not always available in the birth files, which resulted in missing data for some subjects, we evaluated the number of days in the NICU and under mechanical ventilation. The correlation was high between the number of days spent in the NICU and the number of painful procedures  see . Therefon = 18) because the medical files of subjects born before 1992 did not explicitly include a detailed report of the painful procedures . Nonetheless, there is a strong significant positive correlation  between the number of days spent in the NICU and the number of painful procedures for these subjects. It would be unlikely that the subjects for which the files were not complete show a different correlation.Data was not available for all subjects /HRbaseline). We reported HR reactivity for the mean of every 15 seconds or as a mean of the entire 2-minute immersion.Electrocardiogram (ECG) activity was recorded using a standard 3-lead montage sampled at a frequency of 1000 Hz with the Powerlab system and Chart software . ECGs were recorded for two minutes prior to testing (baseline) and during the CPT (immersion). Instantaneous RR intervals were obtained from the ECG waveform with a peak detection algorithm to detect successive R-waves. HR variability analyses in the frequency domain were done subsequently. Fast Fourier transforms were used to calculate the power spectral density of HR oscillations (window length: 1024). The FFT method can provide a noisy estimator of the power spectrum. This is why we used it in combination with Welch's method. This allows us to achieve a consistent estimator by averaging periodogram from overlapping intervals . In our Demographic information was collected prior to testing for each participant. Pediatric ECG electrodes were first installed and baseline ECG was obtained as the participants were asked to relax. Participants were then asked to immerse their right arm in cold water, while ECG was recorded.P < 0.05 (two-tailed) was considered statistically significant.Descriptive statistics are presented as mean ± standard deviation (SD). After checking for assumptions of normality and homogeneity of variances, Kruskal-Wallis tests and 2 × 2 ANCOVAs were used. Bonferroni post hoc analyses were used to evaluate the differences between the specific groups. For the repeated measures of HR during the cold water test, we used the mixed procedure of the SAS System for Windows, version 9.1 . The fixed part of the model contains a linear and a quadratic effect of time and the interaction between each of those factors and the group (preterm or full-term) to test if the linear and quadratic effect changes between groups. To account for the correlated residuals within subjects, a repeated statement with an unstructured covariance structure was used on time. The unstructured covariance structure gave the best Akaike Information Criterion (AIC) compared to other structures  and CS) and therefore was used for the analysis. t = −1.43, P = 0.16). Both groups had a comparable change in HR at the beginning (first 15 seconds) of the procedure . However, an analysis of the change in HR during the CPT, evaluated every 15 seconds, showed that full-term subjects experienced a more stable increase in HR during the immersion compared to the preterm participants. Preterm subjects had a strong increase at the beginning of the immersion and HR slowly decreased over time. An analysis revealed that the birth type (preterm versus full-term) had an impact on the course of the percentage change during the procedure , with age having a significant effect  when entered as a covariate (There were no significant group differences (preterm versus full-term) in resting heart rate (ovariate .There was a different quadratic relationship between the percentage change and time in full-term compared to preterm subjects. The full-term group shows an increase of their percent change in HR over time followed by a decreased (concave function) while the preterm group shows an opposite effect. They show a high increase at the beginning that instantly decreases (convex function).Ps < 0.05) (Ps < 0.03). The preterm groups also differed from one another for those variables (both Ps < 0.001). The high-pain preterm group had more days of ventilation (P < 0.001) and spent more days in the NICU (P < 0.001) than the low-pain preterm subjects. The full-term surgery and high-pain preterm groups differed only in GA, birth weight, and time spent in NICU   according to their type of birth and the length of stay in the NICU and number of days under mechanical ventilation, as proxies for the pain suffered at birth. There were significant differences for GA, birth weight, 5 min APGAR, days under mechanical ventilation, and days spent in NICU between the groups  . Bonferr< 0.001) .P = 0.041) and age at testing was a significant factor (P < 0.001). Full-term surgery and high-pain preterm groups showed significant higher resting HR (Tables A 2 × 2 ANCOVA was used to determine the impact of type of birth and pain experienced at birth on mean baseline HR with age at testing as a covariate. Pain experienced at birth had a significant impact (R Tables  and 4.P = 0.013), HF (P = 0.003), and the ln of the ratio LF/HF (P = 0.004) , with age at testing being a significant covariate (P = 0.027)  who experienced many painful interventions at birth exhibited a complete absence of rise in HR during CPT [Similar results are found when comparing mean percentage change in HR. In fact, high-pain preterm subjects show a lower mean percentage change in their HR. In addition, we found that full-term subjects with surgery also showed a lower mean percentage change in HR. These findings suggest that having suffered intense pain at birth regardless of gestational age alters heart rate changes to experimental pain. To our knowledge, no other study has clearly exposed this relationship. Previous studies only compared preterm to full-term subjects. Preterm infants who experienced a higher number of painful procedures following birth showed less autonomic reactivity , 8. TheyA higher baseline HR was observed in subjects who underwent painful procedures at birth, when age at testing was entered as a covariate in the analysis. This difference is not found if we compare full-term to preterm birth, suggesting that the pain suffered is the cause of the altered HR. This is consistent with previous studies that found that the high-pain preterm group showed a higher resting HR, suggesting a state of chronic arousal , 25, 26.We also found important autonomic reactivity differences in our groups. In fact, full-term surgery and high-pain preterm participants showed a stronger sympathetic activity (LF) and a lower cardiac vagal activity (HF) at rest, suggesting again a state of chronic arousal. Our study confirmed a sympathovagal imbalance in subjects who experienced pain at birth (either born preterm or full-term). This sympathovagal imbalance has also been found in subjects with migraine and alterations in autonomic nervous system function and is thought to contribute to the development of symptoms in these patients . Becausen = 16, low-pain preterm: n = 15, high-pain preterm: n = 16, and full-term surgery: n = 3). Even when 2-minute recordings are used, the stationarity of the signal may be not respected and the Fast Fourier Transform algorithm used to analyse the HRV may lead to misleading results. Therefore, we need to be cautious in the interpretation of the results. Also, we used the length of stay in the NICU and the number of days under mechanical ventilation as proxies for the extent of pain experienced at birth, because the exact number of painful procedures was not always available due to archive procedures. Nonetheless, we were able to determine that the length of stay was in fact correlated to the number of painful procedures experienced at birth in the preterm group. However, other confounding factors like maternal separation or stress could be associated with a longer stay in the NICU and a higher number of days under mechanical ventilation and have an impact on the cardiac pain response. The health of the infant at birth, which could have also increased the NICU stay, as well as stress caused by painful procedures and maternal separation [However, some limitations should be taken into account concerning our study. First, not all subjects were able to complete the entire 2-minute CPT procedure (47%). A minimum recording of two minutes of the heart rate has been shown to be necessary to perform a short-term HRV analysis . Consequparation , could hClinically, our results have important implications because they demonstrate an impact of the length of stay in the NICU and the number of days under mechanical ventilation  on resting and stressed autonomic cardiac activity years after the stay in the NICU. Subjects with prior pain exposure, either born preterm or full-term, showed a higher baseline HR, a stronger sympathetic activity, and a lower cardiac vagal activity even at adulthood. This study adds to the prevailing literature in the sense that the impact on heart rate comes from pain experienced at birth , independently of the birth status (preterm versus full-term), and is still found in young adulthood. This could lead to impaired reactions to pain or stress in later life. It is important to consider the long-term impact of pain at birth, even in the presence of some analgesia in full-term neonates who underwent surgery. We showed that the regulation of the cardiac autonomic system is still altered years after all procedural pain is over. Conclusions of our study show the importance of providing better analgesic care for the neonates and limiting as much as possible painful medical procedures during this critical time.


Elephas maximus) inhabit regions with seasonal climate, but females can still experience 16-week reproductive cycles throughout the year. Whether female elephants nevertheless concentrate births on periods with maximum offspring survival prospects remains unknown. We investigated the seasonal timing of births, and effects of birth month on short- and long-term mortality of Asian elephants, using a unique demographic data set of 2350 semicaptive, longitudinally monitored logging elephants from Myanmar experiencing seasonal variation in both workload and environmental conditions. Our results show variation in birth rate across the year, with 41% of births occurring between December and March. This corresponds to the cool, dry period and the beginning of the hot season, and to conceptions occurring during the resting, nonlogging period between February and June. Giving birth during the peak December to March period improves offspring survival, as the odds for survival between age 1 and 5 years are 44% higher for individuals born during the high birth rate period than those conceived during working months. Our results suggest that seasonal conditions, most likely maternal workload and/or climate, limit conception rate and calf survival in this population through effects on maternal stress, estrus cycles, or access to mates. This has implications for improving the birth rate and infant survival in captive populations by limiting workload of females of reproductive age. As working populations are currently unsustainable and supplemented through the capture of wild elephants, it is imperative to the conservation of Asian elephants to understand and alleviate the effects of seasonal conditions on vital rates in the working population in order to reduce the pressure for further capture from the wild.In seasonal environments, many species concentrate their reproduction in the time of year most likely to maximize offspring survival. Asian elephants ( Cervus elaphus) all females give birth within a 2-month time period from mid-May to mid-July     Wallis . Season ) Wallis . However) Wallis . These t) Wallis . There i) Wallis .. In summary, previous studies of elephants have not yet fully clarified the relationship between birth timing, offspring mortality and seasonal variation in environmental conditions, and in addition they have primarily focused on wild African elephants. African and Asian elephants are known to differ in reproductive endocrinology with a higher rate of noncycling females  experiencing poorer early-life environments showed higher mortality before age 2 years  the distribution of births through the year, and the effects of climate seasonality  on birth timing. We also analyze (2) the short-term effect of birth month on mortality to age 1 month, during which 10% of calf deaths occur  Red List of threatened species IUCN . MyanmarThroughout the year, elephants are not provisioned and forage unsupervised in the forest at night  or whose death could not be confirmed (N = 3) were censored, that is, that they were considered in our analyses until the age for which their survival status is unknown.The Extraction Department of MTE collects and maintains records of all individual elephants. This comprehensive countrywide system is unique to Myanmar and is the equivalent to studbooks kept by Western zoos. Data for each individual include registration number, sex, maternal identity, birth and death dates, location of birth, origin (wild caught or captive born), capture method (if applicable), year of capture and birth dates, identities, and cause of death (if applicable) of offspring. Births of captive-born elephants are recorded precisely. In order to check the health, condition, and working ability of each elephant, individual logbooks are updated by local veterinarians and regional extraction managers at least bimonthly. The multiple sources of data recorded by the MTE  allow effective cross-checking of any apparent errors. Our complete data set collected from the records of the MTE contains details of over 8000 elephants from 11 regions . Each row in the data set was replicated 12 times (one for each possible month) and for each elephant the response variable was “1” for the birth month and “0” for the other 11 months. As such, birth seasonality corresponds to a discrete time event analysis in which row replicates do not introduce pseudoreplication issues , between 1 month and 1 year of age (N = 2114), between age 1 and age 5 years (N = 1887), and between age 5 and age 17 years (N = 866) by birth month using a logistic regression framework. In contrast to the birth seasonality model above, each row of the data set corresponds to a different elephant because only two possible outcomes characterize each elephant . Logistic regression models for each of the four age intervals were implemented as a generalized linear mixed-effect model to predict the probability of death (binary) as a function of three fixed effects: birth month (12 levels), sex (two levels), and birth order (firstborn or later born). We included all two-way interactions between birth month, sex and birth order, and the three-way interaction. We also included three additional random terms to take into account sibship effects , location (11 levels), and cohort effects .We examined mortality in the first month of life . The months in which fewer births occurred than expected were May, June, July, and August . Accordingly, birth rate was significantly associated with the factor month in our time event analysis . We predicted birth rates controlling for sex and birth order, and identified that, similar to raw data, the predicted birth rate was the highest between December and March. Additionally, there was also a significant interaction between month and birth order . The interaction between month and birth order did not differ between sexes as demonstrated by a nonsignificant triple interaction .We found clear evidence that although female Asian elephants can cycle throughout the year, their reproductive rate fluctuates seasonally. Number of births for all timber sites combined varied between months . Both temperature and rainfall significantly predicted the probability of births . A further 5.4% of calves died between ages of 1 month and 1 year, but again there was neither direct association between mortality and birth month, nor significant interactions between birth month and birth order or sex .We then investigated whether the observed seasonal timing of births maximizes immediate calf survival prospects and if mothers are timing their births to the most favorable periods. Overall, 10.0% of calves died during the first month, but neither birth month nor the interaction between birth month and birth order or sex modified this risk significantly , but depending on birth order  with the mortality of firstborn calves being more influenced by seasonality that later borns. Controlling for birth order and other covariates (considered as fixed or random effects), the lowest calf mortality rates predicted from this model corresponded to being born in the months December–April (N = 1173) or not (N = 1177), instead of using the variable birth month with 12 different levels. As such, the overall effect of birth season on probability of death became much clearer . The odds of survival were 1.44 times higher for calves born during the high birth rate period compared to those born during the low birth rate period. A total of 16.9% of those born in the high birth rate period died between the ages of 1–5 compared to 22.7% of those born in the low birth rate period. For the 27.6% of juveniles which died between 5 and 17 years of age, mortality was no longer associated with birth month .Finally, we investigated the long-term consequences of the seasonal timing of birth on mortality risk. A total of 18.1% of calves surviving to age 1 year died before reaching independence at age 5 years. This mortality rate was marginally associated with birth month   or active . On one hand, females may have more opportunities to interact with potential reproductive mates  due to increased time spent in the forest. On the other hand, workload may actively affect reproductive cycling of females, which was shown to be disrupted under poor or stressful conditions (Schneider Another explanation of the observed birth pattern is that it reflects seasonal variation in the number or proportion of pregnancies carried to full term. Even if conception rate is constant through the year, the same maternal and environmental factors described above may be causing an increase in rates of abortion during the work season. However, early-term abortions (known to frequently occur in Asian elephants in zoos; Lueders et al. Our findings have relevance for conservation of Asian elephants because the working population of elephants in Myanmar represents a large population of an endangered large mammal that, despite having considerably better survival rates than Asian elephants in Western zoos (Clubb et al. Therefore, breeding of captive elephants has become increasingly important (Kurt et al.


Small birth size - an indicator of a sub-optimal prenatal environment - and variation in growth after birth have been associated with non-communicable diseases in later life. We tested whether birth size or growth in childhood associated with the risk of hospital admission for alcohol use disorders (AUDs) from early to late adulthood.The sample comprised 6544 men and 6050 women born between 1934 and 1944 in Helsinki, Finland. Data on anthropometric measures were extracted from medical records and diagnoses of AUD from the Finnish Hospital Discharge Register and Causes of Death Register covering a 40-year period from 1969 to 2008.Altogether 171 women (2.8%) and 657 men (10.0%) were diagnosed at a hospital with AUD. After adjusting for major confounders, shorter length at birth, shorter height up to two years of age, and lower weight at two years associated with hospitalization for AUD in women. In men, slower growth in height, particularly from 2 to 7 years, and slower weight gain from 7 to 11 years as well as shorter height and lower weight at 7 and 11 years associated with a diagnosis of AUD in men.Pre- and postnatal growth associates with the risk for AUD later in life differently in women than in men: the fetal period and infancy seem to be the sensitive periods for women, whereas those for men the occur from toddlerhood onwards. In Western European societies, the lifetime prevalence of alcohol dependence is around 5% to 8% with rates four to five times higher for men than for women Within the developmental origins of health and disease framework (DOHaD), the associations between small birth size or slower growth in childhood and health outcomes in later life have been attributed to suboptimal environmental conditions during sensitive time periods that alter the structure and function of key organs such as the brain less frequent alcohol consumption in young adulthood Various non-communicable health conditions, such as hypertension and several psychiatric conditions, show high comorbidity with AUDs Our study aimed to investigate whether length of gestation, body size at birth and physical growth measured serially from birth to 11 years of age associate with hospitalization for or death due to AUDs in the Helsinki Birth Cohort Study (HBCS) sample. In addition to testing for linear effects, we examined whether nonlinear associations existed between birth size and hospitalization for AUDs. We also tested whether associations between physical growth and AUDs stem from comorbid mental disorders.The original study cohort comprised 13345 subjects who were born in Helsinki, Finland at one of the two public maternity hospitals, who visited child welfare clinics, and who received a Finnish personal ID number. Of these participants, we excluded 191 individuals because they moved abroad or died before data collection for the Finnish Hospital Discharge Register (HDR) began in 1969, 226 who were missing data on AUDs, and 334 who were missing data on childhood socioeconomic position (SEP). The cohort available for the analyses comprised 12 594 subjects . The excluded participants had an earlier year of birth (p<0.02), were more often women (p<0.004) and were lighter (p = 0.03) and thinner at birth (p<0.001), but showed no differences from the included participants in childhood SEP, length of gestation, placental weight, or length and head circumference at birth . Furthermore, 847 participants were excluded from the analysis of neonatal characteristics due to missing data on the length of gestation. The Ethics Committee of the National Public Health Institute and the Ethics Committee of the Helsinki and Uusimaa Hospital District approved the HBCS. The study has was conducted in adherence to the Declaration of Helsinki.3) and body mass index (BMI: kg/m2).Data on year of birth, weight (g), length (cm), head circumference (cm), placental weight (g), and date of the mother's last menstrual period were collected from the birth records. Data on monthly changes in weight (kg) and height (cm) from birth to 2 years and annual changes from 2 to 11 years were estimated from child welfare clinic and school health records as described previously Our primary indicator of SEP in childhood was occupational status of the father. These data were derived from school, child welfare clinic, and birth records  and were classified as low SEP , middle SEP (lower middle class: 22.6%), and high SEP (upper middle class: 15.2%). We used highest maternal SEP as secondary indicator of childhood socio-economic environment due to higher number of those belonging to unclassified category (e.g. housewives or students). Of the mothers, 50.9%, 34.2%, 3.8%, 11.1% belonged to low SEP, middle SEP, high SEP, and other/unclassified SEP, respectively Diagnoses were extracted from the HDR, which contained data on all hospitalizations in psychiatric and general hospitals in Finland between 1969 and 2008. The HDR also includes personal and hospital ID numbers, dates of hospital admission and discharge, and primary as well as up to three subsidiary diagnoses at discharge. The HDR is a valid and reliable tool for research Diagnoses were entered into the HDR and CDR according to the International Classification of Diseases, Eighth Revision (ICD-8) until 1986, according to the ICD-9 using the Diagnostic and Statistical Manual of Mental Disorders, Revised Third Edition (DSM-III-R) criteria until 1995, and according to the ICD-10 since 1996. In the current study, the primary diagnoses of alcohol intoxication (ICD-9: 3050A and ICD-10: F10.0) as well as the primary and subsidiary diagnoses of harmful alcohol use, alcohol dependence, and psychotic disorder due to alcohol abuse  from either register served to index the AUDs. In our sample, we identified 746 and 34 AUD cases based solely on the HDR and CDR, respectively, and identified 48 based on both registers .We examined the different AUDs together as one broad diagnostic outcome rather than as separate entities  to make our definition of AUDs more comparable with the definition of AUDs in the DSM-5, that was introduced in 2013. However, in correspondence with ICD-10, our definition of AUDs includes also the primary and subsidiary diagnoses of alcohol psychoses, amnesic syndrome due to alcohol use, and other and unspecified mental and behavioural disorders due to alcohol use, and the primary diagnoses of alcohol intoxification.The primary and up to three subsidiary diagnoses of mental disorders other than those due to alcohol use were drawn with the following codes: 295–302, 304–305, 306.4–306.5, 306.8, 306.98, and 307 from ICD-8; 292, 295–298, 300–302, 304, 3051–3059, 3071A, 3074, 3075A–3075B, 3078A, 3079X, 3090A, 3092C–3099X, and 312 from ICD-9; and F11–F69 from ICD-10 First, we used logistic regression analyses to test whether neonatal characteristics and body size or growth from birth to age 11 were associated with hospitalization for AUDs in adulthood. We tested for nonlinear associations between birth size and hospitalization for AUDs using a model that included a squared term of the variable centered around the grand mean together with a non-squared linear variable. We focused on measurements conducted at birth, at 6 months of age, and at 1, 2, 7, and 11 years of age to represent growth in the prenatal period (at birth), in infancy (up to 2 years), and in childhood (up to 11 years). Physical size measurements served as continuous variables to test for linear and nonlinear associations. All the neonatal characteristics and the measures on attained size at different ages were converted to z-scores by sex. Postnatal growth variables were standardized residuals from linear regression models of weight, height, and BMI where body size at each point in time was regressed on corresponding measures at all earlier points in time, creating completely uncorrelated residuals reflecting growth conditional on history  Although we found no statistically significant interactions between body size or growth by sex , we conducted all the analyses separately for men and women, since previous studies have found major sex differences in the prevalence of AUDs We then tested whether growth conditional on history associated with hospitalization for AUDs. Among men we found that after adjusting for birth year and SEP in childhood, hospitalization for AUDs was associated with slower growth in height from two to seven years  and with slower weight gain from seven to 11 years , while weight gain from two to seven years showed a similar trend . We found no significant associations between growth conditional on history and AUDs in women. Further adjusting for maternal SEP (Model 2) or comorbid mental disorders (Model 3) yielded only one change in significance levels: in men, slower growth in height from two to seven years became non-significant . Restricting the sample to those with mental disorders rendered all significant associations non-significant .We explored whether length of gestation and body size or physical growth up to 11 years of age associated with hospitalization for AUDs in adulthood in 12594 women and men participating in the Helsinki Birth Cohort Study. Different growth periods were associated with increased risk for AUDs in both women and men. After adjusting for gestational age, birth year, and SEP in childhood, shorter length at birth associated linearly with increased risk of hospitalization for AUDs in women. This association was unattributable to maternal SEP or comorbid other mental disorders, although restricting the sample to those with any mental disorders rendered the association non-significant. AUDs showed no association with growth after birth, although women who were hospitalized for AUDs remained on average shorter up to two years of age, and weighed less at two years of age. These results were unattributable to maternal SEP or comorbid mental disorders. In men, birth characteristics showed no association with the risk for AUDs. However, men who were hospitalized because of AUDs had slower growth in height between two and seven years of age; they also gained less weight from seven to 11 years and, as a consequence, were shorter and weighed less at age 7 and 11 years. Similar findings related to weight and height are likely the result of slower growth in height, as there was no relationship with BMI in childhood. In men, maternal SEP or comorbid mental disorders had little effect on these associations, but restricting the sample to those with any mental disorders rendered the associations non-significant.These results are in line with those of previous studies on population-based birth cohorts that have shown an inverse gradient between birth weight and substance-related disorders in young adulthood To the best of our knowledge, this study is the first to explore body size or physical growth in childhood with respect to AUDs in adulthood; consequently, the current findings cannot be directly compared to the previous ones. Cross-sectional studies on associations between body size and psychiatric morbidity show conflicting results, with some linking obesity with psychiatric problems Our finding that restricting the sample to those with mental disorders attenuated many of the associations is in line with earlier findings that smaller birth size associates with a wide range of mental disorders Our results cannot determine the biological and psychosocial mechanisms that underlie the associations between smaller body size and the risk for AUDs. The DOHaD framework postulates that these conditions may relate to environmental psychosocial stress and/or malnutrition that alter the structure and function of the brain in utero and in infancy may have greater effects for women, whereas for men, the sensitive periods occur later, from toddlerhood onwards. Consistent with these results, exposure to prenatal stress associated with larger alterations in monoamine neurotransmission Our findings suggest that sensitive periods with respect to the risk for AUDs may differ for men and women. Exposures The strengths of our study are its longitudinal study design, its relatively large sample of participants derived from the general population, and its use of comprehensive register data. Moreover, we measured growth at several points in time, which permitted us to track critical periods of growth with respect to subsequent AUDs. Our study also has its limitations. First, we have no data on parental mental disorders or risk behaviors such as alcohol use or smoking. Since smoking or alcohol use during pregnancy may increase the risk for small birth size In summary, the associations between anthropometric traits, indicators of the prenatal and childhood nutritional and psychosocial environment, and AUDs showed differences between women and men: shorter length/height at birth and in infancy in girls and in childhood in boys, associated with increased risk for AUDs in later life.


Background: A growing body of evidence indicates a relationship between household indoor air pollution from cooking fires and adverse neonatal outcomes, such as low birth weight (LBW), in resource-poor countries.Objective: We examined the effect of reduced wood smoke exposure in pregnancy on LBW of Guatemalan infants in RESPIRE (Randomized Exposure Study of Pollution Indoors and Respiratory Effects).n = 266) either received a chimney stove (intervention) or continued to cook over an open fire (control). Between October 2002 and December 2004 we weighed 174 eligible infants (69 to mothers who used a chimney stove and 105 to mothers who used an open fire during pregnancy) within 48 hr of birth. Multivariate linear regression and adjusted odds ratios (ORs) were used to estimate differences in birth weight and LBW  associated with chimney-stove versus open-fire use during pregnancy.Methods: Pregnant women , –27 to 204 g] than infants whose mothers used open fires after adjusting for maternal height, diastolic blood pressure, gravidity, and season of birth. The adjusted OR for LBW was 0.74  among infants of stove users compared with open-fire users. Average birth weight was 296 g higher  in infants born during the cold season (after harvest) than in other infants; this unanticipated finding may reflect the role of maternal nutrition on birth weight in an impoverished region.Conclusions: A chimney stove reduced wood smoke exposures and was associated with reduced LBW occurrence. Although not statistically significant, the estimated effect was consistent with previous studies. Low-birth-weight  infants are at high risk for morbidity and mortality in infancy and childhood  as well 2.5 (PM ≤ 2.5 μm in aerodynamic diameter) from wood fires can range from 100 μg/m3 in homes using chimney stoves in good condition to 1,000 μg/m3 in homes using open fires. These concentrations are orders of magnitude higher than those in homes where tobacco cigarettes are smoked, where daily mean concentrations of PM2.5 vary between 25 and 40 μg/m3  and LBW have been described in children in Guatemala , Zimbabwplancha) compared with control families who continued to use traditional open fires  trial. The primary aim of RESPIRE was to examine the incidence of acute lower respiratory infection among children < 18 months of age in households randomized to receive a wood-fueled chimney-stove intervention  in 23 rural Guatemalan communities. Most participants (96%) were members of the Mam-Mayan linguistic group. Households were recruited if they used open wood fires for cooking and had either a pregnant women or a child < 4 months of age. Half of the study homes were randomly selected to receive the chimney stove (intervention group). The other half of the homes continued to use an open fire (control group) until their child reached 18 months, or until the study ended, at which time the family was offered a chimney stove.A total of 266 of the 534 women were enrolled during pregnancy, with the remaining children recruited at < 4 months of age. After enrollment, pregnant women received a baseline pregnancy examination by a physician employed by the study, which included a medical history, urine analysis, blood pressure measurement, and estimate of fetal growth . Fieldworkers visited women weekly in their homes, and women were referred to physicians if the women reported pregnancy complications. Among the enrolled pregnant women, 254 women delivered singleton, live births . Twelve additional births occurred among these women during the study period. Among the 254 women, 240 participated in an ancillary study of adult women’s lung function. In this study, height and nonpregnant weight were measured . Height Once RESPIRE staff were notified of a birth, either by a fieldworker at the weekly home visit or by a family member who contacted project staff, the project manager (A.D.) attempted to visit the home to weigh the newborn within 48 hr. Sometimes practical issues made it difficult to reach the home to carry out measurement, such as delayed notification by family members, minimally skilled birth attendants who did not routinely weigh newborns, and communities located several hours away from the project office. Thus, of the 254 newborns, 208 (81.8%) birth weights were measured within 1 week, including 174 (68.5%) measured within the first 48 hr. Because newborns can lose up to 5–10% of their birth weight in the first week, we restricted our birth weight analysis to the 174 newborns measured within 48 hr of birth. Birth weight was measured in grams using a calibrated Siltec BS1 baby scale with 10-g readability . Infants wore a light shirt provided by project staff to standardize clothing weight. Gestational age was not assessed.Intention-to-treat analysis assumes that those who were randomized to receive the intervention stove had lower exposure to HAP than did those in the control group who continued to use open fires. However, because the RESPIRE trial was designed to monitor infant respiratory illness, many of the pregnant women were recruited in the later stages of pregnancy. They received the stove, which required a 5-week drying period after construction, shortly before, or even a few days after, delivery. Therefore, birth weight outcome analyses were based on actual stove type in use during the pregnancy during the observation period and not based on randomization, resulting in 69 births to women who used a chimney stove during pregnancy and 105 births to women who used an open fire during pregnancy (174 births to 164 women).Personal exposure to CO was used as an indicator of exposure to HAP and as a proxy for PM exposures . PassivePregnant women wore CO tubes for a 48-hr period, with a total of 378 CO tubes worn during 247 (of the 254) pregnancies. Baseline measurements were made before stove construction when all houses were using open fires. During this period, women wore 278 CO tubes during their pregnancy, with 4% of women in the first, 33% in the second, and 63% in the third trimester. During the observation period, CO measurements were made at 3- to 6-month intervals. Pregnant women wore 100 CO tubes during this period, with 6% of women in the first, 14% in the second, and 80% in the third trimesters. If a woman had more than one measurement taken during either the baseline or the observation period, an average of the measures was provided for that period. However, with so few measures available for pregnant women in the first two trimesters with the stove in use, we did not have enough information to incorporate these data into the multivariate birth weight models. Instead, the independent variable is actual stove type used during pregnancy.Statistical methods. We used Stata version 11  for data analyses. Because CO tube data had a right-skewed distribution, both arithmetic and geometric means and respective standard deviations are reported and log-transformed data were used to test for differences between groups. We compared stove groups and birth weight outcomes with Student’s t-test for continuous variables, and chi-square tests of significance for categorical variables. We constructed multivariate linear models of birth weight as a continuous outcome among exposed (open fire) and unexposed (stove) women, based on the stove type that was used during the pregnancy. Covariates were included in the multivariate linear regression if a) there was evidence of a relationship with either outcome (birth weight) or HAP exposure  in the literature, or b) there was an observed statistical association (p < 0.10) with either outcome or exposure using appropriate parametric or nonparametric hypothesis tests. To investigate potential seasonal effects of HAP on birth weight , we examined season of birth in the multivariate model. The three seasons in this region of Guatemala are cold season (15 November to 14 February), warm, dry season (15 February to 30 April), and warm, rainy season (1 May to 14 November). We also examined the effect of season during the three trimesters of fetal growth; interaction terms for season × trimester of stove introduction were included in the regression models. Variables in the model were assessed for multicollinearity using a variance inflation factor (VIF) and were removed if VIF > 10. Outliers were classified using studentized residuals ± 3.0, and a model without outliers was compared with a model that included the “outlier” observations. Leverage plots and partial regression plots were examined to assess any influence individual observations might have on model coefficients. Predicted residuals were plotted against a normal distribution and against the fitted values, to assess for functional form and constant variance.p-Values < 0.05 were considered significant.We used the likelihood-ratio test statistic to compare restricted versus full (saturated) models to determine which model best described the data. We clustered multiple observations on the same woman  to derive standard error estimates that would be robust to nonindependent observations. In the final multivariate model, two outlier observations (one each from chimney-stove and open-fire groups) were excluded because they were found to sufficiently influence the homogeneity of the residuals of the data. An adjusted odds ratio (OR) for presence of LBW among open-fire users, compared with chimney-stove users (reference group), was calculated using logistic regression, and CIs were estimated using robust error estimates. Ethical approval. The study received ethical approval from the institutional review boards at the University of California–Berkeley, University of Liverpool, and Universidad del Valle in Guatemala. Trained fieldworkers obtained verbal informed consent from all participating households before the study.Because most women in this region do not seek health care until late in pregnancy, 86 (33.8%) women were in their second trimester and 157 (61.8%) were in their third trimester at time of enrollment. Among all women enrolled during pregnancy, the mothers and fathers in the stove group were 2 years younger on average than the open-fire group . Women ip = 0.72). Seventeen women (7%) self-reported a personal history of anemia, with no significant difference between groups (p = 0.58). Mean systolic and diastolic blood pressures (mean ± SD) were 1–2 mmHg higher in the open-fire  than in the chimney-stove group , but these differences were not statistically significant. Only one woman had borderline hypertension on examination. None of the women had glycosuria, and < 3% had proteinuria or white blood cells  based on initial urine exam, with no difference between the two groups (p = 0.65).Additionally, we found no statistically significant differences in the number of miscarriages  among the 174 births in this analysis were less economic support (p = 0.10), older maternal age (p = 0.07), higher fundal height at baseline pregnancy exam (p = 0.001), and higher systolic and diastolic blood pressure at baseline medical exam . Variables related to LBW among the 174 births , at the p ≤ 0.10 level, were less economic support (p = 0.09), lower maternal height (p = 0.10), increased live births (p = 0.09), lower fundal height (p = 0.007), and birth during the rainy, warm season (p = 0.04). Infant sex was unrelated to exposure (p = 0.91) and outcome (p = 0.45), and inclusion of this covariate did not improve model fit. Although economic support was related to both the exposure and the outcome, it was not included in the final multivariate model because it did not improve model fit.Characteristics that were associated with use of an open fire versus a chimney stove during pregnancy (p = 0.0003).We measured CO exposures in 247 pregnancies during the baseline and observation periods . BecauseThirty-nine (22.4%) of the 174 infants measured < 48 hr were LBW . CompareThe mean estimated birth weight for the 174 infants was 2,755 g . Unadjusted linear regression models of the mean gram difference in infant birth weight by stove type used during pregnancy demonstrated that average birth weight was 68 g higher  among births to mothers who used the chimney stove versus open fire during pregnancy . The estIn the final multivariate model, the relationship between stove type used during pregnancy and birth weight was not statistically significant. However, maternal height, parity, and season of birth were statistically significant predictors of birth weight, and maternal diastolic blood pressure approached statistical significance. For every 1-cm increase in maternal height, birth weight increased 17 g  after controlling for stove type, gravidity, maternal diastolic blood pressure, and season of birth. For every past pregnancy, birth weight increased 21 g  after controlling for other variables in the model. If born during the cold season (15 November to 14 February), the infant weighed, on average, 296 g more  than if born during the warm, dry season (15 February to 30 April) or the warm, rainy season (1 May to 14 November), after controlling for other variables in the model. For every 1-mmHg increase in maternal diastolic blood pressure (within the observed range of 40–103 mmHg), infant weight was predicted to increase by 5 g  after controlling for stove type, gravidity, maternal height, and season of birth. Sixteen percent of the variance in birth weight was explained by the predictor variables, and the stove explained only 2% of this variance, when we held all of the other variables constant. Although we made all attempts to reduce the root mean square error in modeling the birth weight outcome, the best reduction achievable was 355 g, which explains the wide 95% CIs around the estimates.n = 105), infants born to mothers who used the stove during the first and second trimesters weighed 328 g more , and infants born to mothers who started using the stove in the third trimester weighed 106 g more , after controlling for maternal height, gravidity, maternal diastolic blood pressure, and season. In the subgroup of 69 women who had a chimney stove in use during pregnancy, 52 (75%) had a stove ready for use in the third trimester , 15 (22%) in the second trimester , and 2 (3%) for the entire pregnancy. Among the women who had the chimney stove in the first or second trimester, five (29.4%) had LBW infants; among women who had the chimney stove in the third trimester, eight (15.4%) had LBW infants.Trimester when the stove was introduced was not significantly associated with percentage LBW or gram difference in birth weight between the two groups. Compared with infants born to mothers who used open fires during the entire pregnancy  89-g increase  in birth weight among infants born to mothers using wood-fueled chimney stoves compared with those born to mothers using open fires. These results are similar to those of four other studies that assessed the relationship between birth weight and maternal exposures to HAP . Unlike p = 0.05) more, on average, whereas infants born to women using wood-fueled improved stoves were 32 g heavier, on average, than were those born to women using open fires. Some of the improved stoves in that study were poorly maintained or repaired, which may have biased the estimate toward the null. A study of 634 infants in Pakistan found a nonsignificant decrease in birth weight in infants born to women using wood fuel compared with infants born to natural gas users   weighed 175 g more  than did infants born to mothers who used wood, straw, and dung as fuel . In a st to 9 g) . AlthougDespite limitations of a small sample size, possible birth weight measurement error due to delay between time of birth and time of measurement, exposure misclassification, other unmeasured exposures to wood smoke , and air pollution reductions primarily restricted to the third trimester, we identified a potentially clinically significant, but not statistically significant, difference in birth weight. Because of possible exposure misclassification among women who received their stove days before birth, we created an exposure variable indicating whether women had a stove in use for at least 60 days before delivery. This reduced the stove group from 69 to 34 births, introduced uncertainty into the model, and thus reduced our power to detect an effect from stove type with an estimated 63 g increase . Because of the heavy demands on field personnel to meet the primary aims of the RESPIRE trial, we were not able to assess newborn gestational age at the home deliveries and are thus not able to state what proportion of LBW was due to preterm birth. In 2009, the primary author trained 10 local traditional birth attendants who work in the communities that previously participated in RESPIRE. Using a cell-phone notification system, traditional birth attendants notified us of home deliveries. Within a 2-week period, we evaluated within 48 hr of birth 22 live newborns who were delivered at home. Preliminary analysis of estimated mean gestational age assessment using the New Ballard scale  was 36.1A further limitation of the present study was the small number of maternal CO exposure measurements. A recent systematic review of 41 studies in industrialized countries examined the effect of a range of ambient air pollutants on adverse birth outcomes; 13 studies looked at CO and LBW . Although averaging times and exposure during pregnancy intervals varied across the studies, significant estimated effects on birth weight were reported for CO as low as 1–2 ppm (of continuous exposure) in the third trimester based on in single-pollutant ambient models . These l2 for maternal undernutrition. This is comparable with the 2002 Guatemalan maternal–child health national survey, which found maternal undernutrition to be 1.9% for women of reproductive age (PAHO 2007).Another limitation is the lack of information on diet and maternal weight gain during pregnancy. Chronic undernutrition was marked among the participating mothers and is reflected in their short stature. In this study, 2.0% of nonpregnant women were below the standard reference cutoff of 18.5 kg/mOur study has several strengths. This study of HAP exposure and birth weight is the first to explore a trimester effect with the use of improved stoves, although few women were observed starting from the first or the second trimester. Differential exposure was limited primarily to the third trimester and after stove construction. The velocity of fetal growth is largest during the third trimester, so limiting smoke exposures during this period is crucial . We obseA further strength of our study is the low prevalence of active tobacco smoke and ETS exposure. In other studies measuring the impact of solid fuels used for cooking, LBW could potentially be attributed in part to unmeasured tobacco exposures if smoking is common. In Pakistan, Weekly visits made by trained fieldworkers ensured that the chimney stove was functioning as intended. During the RESPIRE trial, the chimney stove reduced kitchen air pollution by approximately 90%. Among all 529 women (pregnant and not pregnant) monitored during the 2-year trial, arithmetic mean personal exposures over 48 hr were reduced by 54%, from 4.8 ± 3.6 ppm to 2.2 ± 2.6 ppm, in the chimney-stove group . Among ptemazcal, Mam chuj). Temazcal use could have a much greater effect on LBW than kitchen exposures. Data presented here did not incorporate these significantly elevated exposures, because women were asked to remove their CO tubes before entering the temazcal to bathe, because high-humidity conditions would interfere with the accuracy of the tube measurement. These extreme, acute CO exposure levels affect the entire population of temazcal users and could contribute to the high incidence of LBW observed in the RESPIRE population .In these communities, there are other important sources of smoke exposure, and certainly there are exposures to other copollutants in wood smoke besides CO. Roughly 85% of the population uses a wood-fired steam bath for 30–45 min several times a week . It is possible that higher birth weight during the cold season is attributable to increased food availability during the harvest period, which occurs in the months preceding the cold season. Increased fetal weight may be attributable to improved nutrition during the second trimester and third trimester, when fetal weight gain accelerates, although this cannot be verified from our data. A final strength of this study is the intensive, weekly household surveillance that was employed during RESPIRE. This increased our ability to weigh infants born at home soon after birth. Our LBW prevalence was 22.4%, almost double the national reported prevalence of 12% (PAHO 2007). The national LBW rate includes urban populations, typically characterized by higher socioeconomic status, better access to medical care, and cleaner cooking fuels.To the best of our knowledge, this is the first study to monitor stove use on a weekly basis, measure CO exposures during pregnancy, and explore the impact of wood smoke, by trimester of exposure, on birth weight. Although we randomized pregnant women to a stove intervention during RESPIRE, most of the women were in their third trimester; given the timing of stove allocation, an intention-to-treat analysis was not possible. Although the chimney stove reduced exposure to HAP, exposures remained high for many women and infants. Exposure reduction through well-operating chimney stoves would not only improve the quality of life of the cook and her family but also reduce LBW prevalence, based on evidence from ambient and HAP studies. Future research is needed to quantify women’s personal exposure to HAP during each trimester of pregnancy to assess the effect on adverse newborn outcomes, including birth weight and preterm birth. This would provide supportive evidence to encourage delivery of improved stoves and other clean energy technologies to millions of women and their families in low-income countries who continue to burn solid fuels for cooking and heating.