Recent steep declines in child mortality have been attributed in part to increased use of contraceptives and the resulting change in fertility behaviour, including an increase in the time between births. Previous observational studies have documented strong associations between short birth spacing and an increase in the risk of neonatal, infant, and under-five mortality, compared to births with longer preceding birth intervals. In this analysis, we compare two methods to estimate the association between short birth intervals and mortality risk to better inform modelling efforts linking family planning and mortality in children.Our goal was to estimate the mortality risk for neonates, infants, and young children by preceding birth space using household survey data, controlling for mother-level factors and to compare the results to those from previous analyses with survey data.We assessed the potential for confounding when estimating the relative mortality risk by preceding birth interval and estimated mortality risk by birth interval in four categories: less than 18 months, 18–23 months, 24–35 months, and 36 months or longer. We estimated the relative risks among women who were 35 and older at the time of the survey with two methods: in a Cox proportional hazards regression adjusting for potential confounders and also by stratifying Cox regression by mother, to control for all factors that remain constant over a woman's childbearing years. We estimated the overall effects for birth spacing in a meta-analysis with random survey effects.We identified several factors known for their associations with neonatal, infant, and child mortality that are also associated with preceding birth interval. When estimating the effect of birth spacing on mortality, we found that regression adjustment for these factors does not substantially change the risk ratio for short birth intervals compared to an unadjusted mortality ratio. For birth intervals less than 18 months, standard regression adjustment for confounding factors estimated a risk ratio for neonatal mortality of 2.28 . This same effect estimated within mother is 1.57 , a decline of almost one-third in the effect on neonatal mortality.Neonatal, infant, and child mortality are strongly and significantly related to preceding birth interval, where births within a short interval of time after the previous birth have increased mortality. Previous analyses have demonstrated this relationship on average across all births; however, women who have short spaces between births are different from women with long spaces. Among women 35 years and older where a comparison of birth spaces within mother is possible, we find a much reduced although still significant effect of short birth spaces on child mortality. Previous analyses have demonstrated increased neonatal, infant, and child mortality with short preceding birth spaces. This relationship has been estimated as an average or adjusted average increased risk across all births; however, women who have short spaces between births tend to be different from women with longer spaces. Among those instances where a comparison of birth spaces within mother was possible, we estimated the effect of short birth spaces on these mortalities and found a reduced, although still significant, effect.There is an extensive literature linking family planning programmes, contraceptive use, fertility, and the health of infants and young children. Many studies have found a strong association between increasing use of contraceptives or reduced fertility and decreases in child mortality. One explanation of this relationship has been that as women have access to contraceptives, their children are born in different circumstances more favourable for survival .Circumstances at birth conferring excess risk that also tend to decline with increasing contraceptive use have been broadly defined by three categories: age of mother (young and old age), birth parity (first births and higher parity births), and short intervals of time between births or pregnancies (less than 24 or 18 months between births). There has been considerable effort to quantify the risk associated with these factors. In general, analyses to estimate relative risk have proceeded by comparing birth outcomes among various groups with observational data. These different birth conditions have been repeatedly linked to increased poor birth outcomes  using information from trial results and analysis of facility-based data , but also some Multiple Indicator Cluster Surveys (MICS)] to link birth conditions and poor maternal and child health. Surveys have reasonable accuracy when it comes to their recorded birth histories and the real timing of births . HoweverThe association between these factors and increased child mortality is not necessarily causal or attributable. Associations could be confounded by other factors, such as differences between mothers who have births with these conditions or circumstances that cause both short interbirth intervals and child mortality, such as limited access to health services, including contraception. In addition, an association could also be explained by women being more likely to use contraceptives and reduce fertility when child mortality has dropped (replacement effect) .The relationship between birth spacing and mortality is of interest for determining the attributable fraction of child mortality and predicting the impact of related interventions with the Lives Saved Tool (LiST) . AnalyseWhile in theory this approach eliminates some confounding, such adjustment does not necessarily rule out all possible bias due to related factors. First, there may be confounders that have not been measured or have been poorly measured, for example, access to health services or women's empowerment. Second, even confounders that are known and measured may violate model assumptions and lead to biased or imprecise estimates of effect .Recent work by Kozuki and Walker  used a dIn this paper, we have two primary purposes. First, we investigate the background differences between women who have short birth intervals and those women who do not. Second, we directly compare the standard cross-sectional analysis of mortality risk to the within-mother analysis.We used complete birth histories from 145 standard, interim, or continuous DHS household surveys conducted since 1998, including DHS phases III through VI, from 66 unique countries. We included retrospective birth histories from all of these surveys in our analysis of short birth intervals and the women who have children with short birth intervals, while excluding multiple births.We used these birth histories to assess the potential for confounding in the relationship between short birth intervals and child mortality. To describe women with short birth intervals, we grouped mothers according to their shortest birth interval in four categories  and summarised the wealth quintile, education, age, and fertility of each group. We also described the observed mortality rates for the children of these women in each category of preceding birth interval.In addition to assessing the potential for confounding, we also analysed these birth history data to assess mortality risk. For this primary analysis, we also excluded birth histories from women who were younger than 35 when their birth history was recorded, at the time of survey. We limited our analysis to women who were older at survey for several reasons. First, these women are nearing the end of their childbearing years, and so we can estimate their fertility, an additional potential confounder for the effect of birth spacing on mortality . Second,We analysed the relationship between birth spacing and neonatal, infant, and under-five mortality using two approaches. First, we estimated the mortality risk ratio by interbirth interval using a standard regression adjustment for potential confounding factors. We used Cox proportional hazards for mortality  outcomes based on retrospective birth histories, censored at the appropriate age and including mother's education in three categories, along with wealth quintile, partner's education, need for family planning satisfied, completed fertility, and area of residence  as potential confounders in the regression analysis, described in In addition to this first approach, we estimated the mortality risk ratio by interbirth interval in a within-mother comparison. We also used Cox proportional hazards for mortality outcomes; however, we stratified by mother to compare mortality only for women who had variety in their interbirth spacing. This second analysis was similar to conditional logistic regression, where mortality outcomes are compared for different outcomes within woman, except that conditional logistic regression estimates odds ratios, while Cox regression estimates mortality rate ratios , 15. We Stratifying proportional hazards regression matches births within woman and so strictly controls for all factors that are constant for mothers over their childbearing years . There iWe estimated the effect of birth spacing separately for each survey using these two methods. We then combined these results from multiple surveys in a meta-analysis with random survey effects, weighted by the estimated standard error of the birth spacing effect for each survey, for an overall effect estimate for each method. In addition, we assessed the potential for confounding in the within-mother analysis by examining the association between birth spacing and parity. We also described those births that did not contribute to the second within-mother analysis.These analyses reveal the differences that exist between mothers who have short-spaced births compared to those who do not. In general, the women who had at least one birth that came 18 months or less after a preceding birth had lower education, were more often from the poorest wealth quintile, and had higher levels of fertility. These factors, and perhaps others that are unmeasured, have the potential to confound the association between shorter birth intervals and risk of mortality in the children of these women.We first compared the crude mortality ratios to the adjusted ratios for both the standard regression adjustment approach and the within-mother analysis using the stratified Cox proportional hazards regression. As can be seen in In Matching births within-mother controls completely for factors that are constant over a woman's childbearing years. Education and socio-economic status, for example, would not be expected to change substantially during this time. Other factors identified for their relationship with child mortality are known to vary significantly during this time period.There are two possible selection issues that might make the within-mother comparison biased. First, if there is a difference in parity between the short-spaced and regular-spaced births, it could confound the effect of short birth spacing with the association of mortality and parity. To test this possibility, we computed the average parity for each birth in the within-mother analysis for the various interbirth intervals. Not surprisingly, average parity among all births is highly variable across the DHS in this analysis. However, the difference between average parity of short-spaced births and the average parity of regular-spaced births is very small. The average parity of births with preceding space of 24–35 months was 3.76 (range: 2.3–4.6), while the average parity for births with less than 18 months preceding space is 3.82 (range: 2.3–4.7). The average parity for births with 18–24 months preceding space is 3.76 (range: 2.3–4.5). Scatter plots of these average birth orders are shown in A second possible limitation of the within-mother comparison is that requiring mothers to have both a short- and a regular-spaced birth to contribute to the analysis might result in a biased sample of short-spaced births. This would occur because matching births with short spaces to births with optimal space within mother may exclude some births from analysis, since short-spaced births with no optimally spaced sibling have no potential match and do not contribute to the estimated effect of short birth spacing on mortality. This scenario inevitably arises for some women, for example, women with only two births. Such a woman can only have one birth with an interbirth interval, because first births are not subject to a birth space.In order to assess this possible source of bias, we compared the total number of short-spaced births among women who were at least 35 years old at survey to the number that were used in the within-mother comparison. Of 2.4 million births to women surveyed at 35+years, there were 0.74 million with a preceding space less than 24 months. Overall, 0.69 million (93%) of these births had siblings who were not subject to a short preceding birth interval, and so were matched within-mother and contribute to analysis.While overall the percent of births lost due to matching within mother is low, some surveys were more likely to exclude births than others. The 2007 Ukraine Survey, for example, has the lowest percent of matched short-spaced births (45%), so that more than half of all short-spaced births to women surveyed at 35+ years from this survey were not contributing to analysis. This analysis suggests that overall the within-mother technique seems to provide a strong representation of short-spaced births to women who were 35 or older at the time of survey. The vast majority of surveys found that over 90% of short-spaced births could be matched with a regular-spaced birth. The one limitation is that for countries with low fertility such as Ukraine, the percent of matches is reduced and here the risk ratios estimated by the within-mother comparison may not be representative of all short-spaced births to women surveyed at 35 or older.We estimated the effect of preceding birth interval in four categories on the risk of neonatal, infant, and under-five mortality with recent household survey data from 66 countries. We found ample evidence of a potential for confounding factors when estimating the relationship between birth spacing and mortality in these survey data. We then estimated the relative neonatal, infant, and under-five mortality risks based on two different methods: first, to replicate how this relationship has been determined historically, by including available factors in a proportional hazards regression and, second, with mothers as a unit of stratification among women 35 years and older at survey. Our assumption was that using mothers as strata would be a more efficient method to control for confounders, and the resulting effect estimate would be closer to a direct effect, since this analysis controls for all factors that are constant over a woman's childbearing years. Both this and the former regression adjustment method clearly indicated a statistically significant effect of birth spacing on child mortality. However, the effect estimated by the within-mother analysis was approximately 30% smaller than that estimated with the adjusted regression for neonatal mortality, strongly indicating that there is confounding in the regression-adjusted estimate. Rutstein and Winter  used metWe found a similar effect for our restricted sample with a standard adjustment method  to that found in a similar unrestricted analysis by Rutstein and Winter (We restricted our sample to women who were older at the time of survey, and so some births to young women were excluded. In addition, we were not able to include all short-spaced births in the within-mother analysis, because even among women providing birth histories at 35 years or older, not all births had appropriate matches, that is, births by the same mother that were not short-spaced. This was especially true for surveys or countries with low fertility rates. However, we expect that countries with low fertility are also in general countries with lower neonatal and child mortality. Thus excluding births primarily from low-fertility areas may be biased away from the null to estimate an effect larger than the direct effect. In addition, the overwhelming majority of surveys contributed at least 90% of all short-spaced births reported in the birth histories from older women. Finally, due to the matching and restriction of birth history information to older women, we were unable to generalise our results to the entire sample of interest.An estimate of the causal relation between birth spacing and neonatal, infant, and under-five mortality is of interest for determining the attributable fraction of child mortality and predicting the impact of interventions with the LiST software. A strong and statistically significant effect has consistently been identified of short preceding spaces on increased child mortality. Using matching and restriction, our study indicates an approximately one-third reduction in the risk of neonatal mortality. We present an alternative comparison of preceding birth spaces within women 35 years and older to estimate a still significant but less strong effect.


In this work, we introduce the history and mechanisms of surface enhanced Raman scattering (SERS), discuss various techniques for fabrication of state-of-the-art SERS substrates, and review recent work on robotizing plasmonic nanoparticles, especially, the efforts we made on fabrication, characterization, and robotization of Raman nanosensors by design. Our nanosensors, consisting of tri-layer nanocapsule structures, are ultrasensitive, well reproducible, and can be robotized by either electric or magnetic tweezers. Three applications using such SERS nanosensors were demonstrated, including location predictable detection, single-cell bioanalysis, and tunable molecule release and monitoring. The integration of SERS and nanoelectromechanical system (NEMS) devices is innovative in both device concept and fabrication, and could potentially inspire a new device scheme for various bio-relevant applications. I. I96]. I2 and magnetic nanorod embedment can be readily motorized by electric tweezers . N. N106]. The concentration of molecules for either single or multiple molecular species decreases with time monotonically . This isC) is given by:0C is the concentration of molecules in the bulk liquid. The molecule release rate (k) is proportional to D) and the thickness of the static diffusion layer (ω), we can readily determine the corresponding value of k by fitting the experimental results with Equation (2)  a,b. It inomotors c. The me λ~v−0.5 . Also kn release d. Note tTo the best of our knowledge, the previously discussed mechanism for tuning the molecule release by mechanical rotating nanomotors is the first of its kind. It is achieved from the unique motorized SERS nanosensors that can rotate while simultaneously detect with ultra-sensitivity. Note that molecules that can be tunably released as shown in the demonstration  are not 10, employed photonic devices and near-field effect to further boost the SERS enhancement, and adopted the electric and magnetic tweezers to motorize the SERS nanosensors for location predicable sensing and single live cell analysis. The assembling of the SERS nanosensors into rotary nanomotors enabled tunable molecule release and the time-dependent detection, which is the first of its kind. Overall, the research discussed in this work could open up many new opportunities in the field of biochemical detection, cell-cell communication, and drug delivery.In summary, we have introduced the history and mechanisms of surface enhanced Raman scattering (SERS), discussed the state-of-the-art fabrication techniques of SERS substrates, and presented challenges in this field before reviewing our recent work on robotized SERS nanosensors for chemical detection. We rationally designed and fabricated SERS substrate to realize ultrasensitive and well reproducible molecule detection with an enhancement factor (EF) of ~10


First childbirth in a woman’s life is one of the most important events in her life. It marks a turnaround when she might have to drop roles of career building and education, for motherhood and parenthood. The timing of the commencement of these roles affects the child bearing behavior of women as they progress in their reproductive ages. Prevalent early first childbirth in Nigeria has been reported as the main cause of high population growth and high  fertility, mortality and morbidity among women, but little has been documented on the progression into first birth as well as factors affecting it in Nigeria. This paper modelled timing of first birth among women in Nigeria and determined socio-demographic and other factors affecting its timing.We hypothesized that background characteristics of a woman will influence her progression into having first birth. We developed and fitted a survival analysis model to understand the timing of first birth among women in Nigeria using a national representative 2013 NDHS data. Women with no children were right censored as of the date of the survey. The Kaplan Meier survival function was used to estimate the probabilities of first birth not occurring until certain ages of women while Cox proportional hazard regression was used to model the timing of first births at 5 % significance level.About 75.7 % of the respondents had given birth in the Northern region of Nigerian compared with 63.8 % in the South. Half (50.1 %) of the first childbirth occurred within the 15–19 years age bracket and 38.1 % within 20–29 years. The overall median survival time to first birth was 20 years , 27 years among women with higher education and 18 years for those with no formal education. The adjusted hazard of first birth was higher in the Northern region of Nigeria than in the South , and higher in rural areas than in urban areas . Also, hazard of earlier first birth tripled among women with no education  compared to women with higher education. The significant factors affecting age at first birth are education, place and zone of residence, age at first marriage, religion, ethnicity and use of contraceptives.This study showed that progression into early first birth is most affected by the education standing of women as well as age at first marriage. Delay of first childbirths as a strategy for fertility reduction and maternal health improvement can be achieved if women are empowered early in life with quality education. Stakeholders should therefore, give adequate attention to educating the girl child. Adverse socio-cultural norms of betrothing and marrying young girls should be abrogated, while health education and promotion of need to delay child bearing must be intensified especially among rural dwellers and also in Northern Nigeria. Age of initiation of childbearing among women has been reported to have strong effect on the demographic behavior of women and the entire population , 2. The On the contrary, importance has been attached to delayed age at first child birth. While it reduces the rate of maternal and child morbidity and mortality on one hand, especially in Nigeria where the maternal mortality ratio (MMR) and infant mortality ratio (IMR) are above African Average . On the Factors affecting age at first child are diverse and are of different levels. There are individual factors, family factors, societal factors as well as economic, national and international factors as illustrated in Fig. The most prominent among the influencing factors are the individual and family background. In this study, we hypothesized that individual and family characteristics would influence the timing of first childbirth among women. On educational attainment, it is not unlikely that girls who could not progress in their educational pursuit would consequently start child bearing earlier than others. Thus, girls from the northern part of the country where literacy level has been put at 22 % compared with 80 % in the south , 10 mighRural area is about a two thirds of Nigeria landscape and contributes about a third of the population , 18; thiStudies have identified peer pressure as one of the factors shaping behavior and attitude of adolescents , 15, 16.Ability of females to get children might affect the age at which they begin parenting. Number of miscarriages and stillbirths before first birth can be used as a proxy to how fecund a woman is. A less fecund woman would ultimately have a later age at first birth than those that had no miscarriage before first birth , 15, 20.Cultural practices such as betrothing girls right from birth or as soon as they are about ten years old to a would be husband , 21, canEducation and employment have dual way relationship with age at first birth. Whether decisions by teenagers on entrance into motherhood are deliberate or not, it has consequential effects on their educational attainment and employability. It is extremely difficult and very rare to combine teenage motherhood with education. While determination to acquire higher education might lead to postponement of first child birth on one hand, early first child birth might stop, or at least temporarily halt, educational process on the other hand , 19, 25.Effect of age at marriage may be very significant on fertility and the age at which women began childbearing  as it isGlobally, age at first child birth is been delayed , 5, 27; While developed countries such as the USA have consistently reported high and increasing age at first birth, rising from 22.6 to 25.3 years, a 12 % increase between 1980 to 2010 . AlthougData from the 2013 Nigeria Demographic and Health Survey (NDHS) [We extracted information on the women’s background characteristics, sexual and reproductive history and knowledge, source and use of contraception. The dependent variable in this study was age at first child birth while region and geographical zones of residence, education, religion, residence and ethnicity were the independent variables. Included also as independent variables are responses on; if the woman ever smoked, whether she had terminated a pregnancy or not and whether she has ever used something to prevent pregnancy. We collapsed the six zones into two regions: The North Central, North East and North West constituted the “North” while South East, South South and South West formed the “South”. We used women’s current educational attainment as a proxy for education as of the time of first child birth. This is justified, because the educational status does not change for persons who had none or primary education throughout their lifetime since primary education is mostly attained at age 12. “Ever smoked” was used as a proxy for peer pressure. We used survival analysis to model the determinants of age at first birth.Survival analysis is analysis of history of events which uses statistical procedures to deal with analysis of time duration, until one or more events of interest happen. It is usual in a follow up study, such as the current study, for some participants not to have experienced the event of interest at the end of the study or some participants were “lost to follow up” or some might have withdrawn during the study. Bias may be introduced if these categories of participants were excluded in further analysis as they could possess unique characteristics that could be useful in answering the research question. In such cases, the length of time the participants stayed in the study would be recorded as their study time and marked as “censored”.S(t) and hazard function h(t). In relation to the present study, the survivor function gives the probability that a woman “survives” longer than some specified time t without a birth, while the hazard function gives the instantaneous potential per unit time to have a first childbirth after time t, given that the individual had not had a first childbirth up to time t. Survival and hazard function are mathematically denoted byTwo quantitative terms are important in survival analysis. They are the survivor function andS(t)) which describes the probability of not failing before time t, hazard function (h(t)) addresses the failure rate at time t among those individuals who are alive at time t. Also two variables are compulsory in survival analysis; they are survival time and the censoring index. The “survival time” or “follow-up time”, is assumed to be a discrete random variable that takes on only positive integer.In contrast to the survivor function  were obtained from equation  is the baseline hazard function, xi are the covariates and βi are the coefficients.We applied the Cox-proportional Hazard model to the age at first birth. The model assumed that proportion of hazards are constant from time to time. In proportional hazard model, the effect of a unit increase in covariate is multiplicative with respect to hazard rate. The Cox model gives an expression for the hazard at time t for a subject in group is assumed to beWe also determined Cox regression estimates for all levels of each of the covariates. In which case, the hazard at time The coefficients are assumed to be the same, regardless of group, but the baseline hazard can be group specific. The sign of the coefficient indicates how a covariate affects the hazard rate. The hazard ratio (HR), expressed as the exponentials of the coefficients, implies more exposure to event of interest if >1, HR < 1 means low exposure while HR = 1 has no effect on the exposure. The statistical significance of the coefficient indicates whether these changes in the expected duration will be statistically significant or not. In the stratified Cox analysis, we tested whether the proportional-hazards assumption was violated using the significance of the hazard ratios, the log likelihood tests and the Wald chi square statistics.p-value = 0.05. We used the Stata (version 13) statistical analysis software for all the analysis.The significant variables in the independent Cox regression were plugged into the multiple Cox regression so as to control for the effects of other variables. We fitted four models. The first model involved the two most significant independent variables at the bivariate logistic level, Model II involved the other socio-cultural characteristics of the women . In model III we added education and age at marriage to the socio-cultural factors while model IV is the full model. The log likelihood tests and the Wald chi square statistics were used to select the best model. The data was weighted to adjust for differences in population in each state and FCT. Statistical significance was determined at About 70.9 % of the 38,948 respondents had given birth to at least a child, 75.7 % in the North compared with 63.8 % in the South, 86.4 % among women with no education and 55.3 % among women with higher Education. Half 50.1 %) had their first childbirth within the 15 to 19 years age bracket, 38.1 % within age 20–29 years, 9.0 % before attaining age 15 and 2.8 % at 30 years or later. Almost 70 % of Northern women had given birth before age 20 years compared with 43.0 % in the South as shown in Table  % had thThe overall median age at first birth was 19 years, 18 years in the North and 20 years in the South; 20 years in urban areas and 18 years in the rural areas; 17 years among Hausa/Fulanis and Muslims compared with 21 years among Yorubas and Igbos. The median age at first birth was 24 years among women with higher education compared with 17 years, among those with no education  given that the woman has had no child before that time. The IR was highest (0.048) among women with no education, who married before age 15 (0.059), from North West (0.045), Hausa/Fulani (0.045), had only primary education (0.044) and lives in North East (0.041). The hazard of early first birth was higher in all zones than in the South West except in the South East. The hazard of first birth in North East and North West was 1.86(95 % CI: 1.79-1.94) and 2.26(95 % CI: 2.17-2.35) respectively, higher than hazard of their counterparts from the South West. The hazard of early first birth in rural areas almost doubled, 1.72(95 % CI:1.68-1.76), the hazard among those from urban areas. Also, hazard of first birth among women with no education was over four times higher, 4.43(95 % CI: 4.22-4.64) than among women with higher education as shown in Table The overall Median Survival Time (MST) was 20 years, with 19 years in the Northern region and 22 years in the south. Also, MST was 27 years among women with higher education and 18 years among those who never had education, 24 years in the South East and 18 years in each of North West and North East. The overall incidence rate (IR), technically defined as the probability that an event will occur at time The survival functions of the age at time of first childbirth by respondents’ characteristics are shown in Fig. p < .05; model I) and among those with no education  and primary education . Model II contained only demographic characteristics without education; this was to see possibility of having significant factors predicting age at first childbirth besides education. Hazards of age at first childbirth were higher in North East  and North West  than in South West but lower in North Central and South South. Hazard of first child birth was higher  than in urban area and higher  among the Hausa/Fulanis, than any other tribe.In Table p < .05; model IV and North East was 1.26; p < .05; model IV compared with South West. Hazard of early first birth was higher in the rural areas ) compared with women in Urban areas and but lower aHR = 0.79; p < .05; model IV) among women who have never used anything to prevent pregnancy compared with those who did. Similarly, women who had ever had pregnancy terminated had higher hazard  than those who never did. The full Model (Model IV) was the best model as it had the best statistics  [The data used for this study is readily available with DHS Measure.


We report the first case of the successful detection of postpartum unruptured vaginal pseudoaneurysm using power- and pulsed-Doppler ultrasonography after delivery.A 38-year-old primiparous Japanese woman had a vaginal laceration with pulsatile bleeding after delivering by vacuum extraction. Transvaginal ultrasonography of the repaired vaginal wall showed an 18 × 20 mm hematoma within which a 6 × 7 mm pulsating anechoic mass was detected. Power-Doppler ultrasonography showed blood flow signals in the anechoic mass. Arterial waveforms detected in pulsed-Doppler mode were suggestive of unruptured pseudoaneurysm. Careful monitoring with contrast-enhanced computed tomography showed an increase in the size of the pseudoaneurysm on the fourth postpartum day. On the sixth postpartum day, massive vaginal bleeding occurred. Emergency angiography revealed strong staining with extravasation from the left vaginal artery, confirming the diagnosis of pseudoaneurysm. Embolization for hemostasis was successfully performed.As far as we know, our case is the first in which an unruptured vaginal pseudoaneurysm was diagnosed using ultrasonography. The differential diagnoses of pseudoaneurysm are arteriovenous malformations including arteriovenous fistula. This case had the typical ultrasonographic patterns of pseudoaneurysm in which the presence of one or two cystic masses in B-mode and color- and/or power-Doppler flow signals was demonstrated along with high-resistance arterial flow waveforms in pulsed-Doppler mode. Sequential examinations of contrast-enhanced CT showed ongoing development of the pseudoaneurysm. In retrospect, we could have performed angiography for embolization when the unruptured pseudoaneurysm was diagnosed, or at the latest when ongoing development of the pseudo-aneurysm was recognized, irrespective of whether symptoms were present.Ultrasonography is a non-invasive and clinically useful modality in the differential diagnosis of pseudoaneurysm. Contrast-enhanced computed tomography with or without ultrasonography can be useful for sequential monitoring of the size of unruptured pseudoaneurysms. Pseudoaneurysms arise from disruption of arterial wall continuity by inflammation or events such as vascular trauma. The pseudoaneurysm margins are formed by the thrombus originating from the injured artery and are not surrounded by three arterial layers. Pseudoaneurysm rupture can lead to devastating patient outcomes, including life-threatening bleeding. Ultrasonography is a useful modality in the diagnosis of unruptured pseudoaneurysm occurring after cesarean section or D&C , including arteriovenous fistula, and true aneurysm. Ultrasound diagnosis of pseudoaneurysm is based on the presence of one or two cystic masses with color- and/or power-Doppler flow signals and high-resistance arterial flow waveforms in pulsed-Doppler mode, often in association with a to-and-fro waveform pattern (Nagayama et al.We did not initially perform embolization of the unruptured pseudoaneurysm in this case, because the patient did not have clinical signs, including bleeding from the vaginal laceration. Instead, we carefully monitored the size of the pseudoaneurysm using ultrasonography and contrast-enhanced CT. Sequential ultrasound examinations did not show growth, but contrast-enhanced CTs showed ongoing development of the pseudoaneurysm. Ultrasonography scans various sections of a targeted mass and an image of the maximal dimension can be viewed. In contrast, CT scans the targeted mass in 5-mm slices perpendicular to the body axis and the image can be viewed as an average within the sliced area. In theory, both contrast-enhanced CT and ultrasonography can detect changes in the size of pseudoaneurysms once they grow beyond the resolution. One possible reason that ultrasonography may fail to detect changes in pseudoaneurysm size is improper scanning.The definitive diagnosis of pseudoaneurysm is made by angiography. However, ultrasonography is a non-invasive and clinically useful modality in the differential diagnosis of pseudoaneurysm. Contrast-enhanced CT with or without ultrasonography can be useful for sequential monitoring of the size of unruptured pseudoaneurysms.The average size of vaginal pseudoaneurysms including our case is 11 mm, with a minimum size of 6 mm (Matsuhashi et al.Vaginal pseudoaneurysm is a rare complication after delivery but there may be unrecognized cases that spontaneously resolve. Routine transvaginal ultrasound examinations using B-mode may be clinically useful after severe vaginal laceration to rule out small and medium-sized hematomas, which may be missed on digital examination. In patients with severe vaginal laceration with pulsatile bleeding and in those with hematoma detected on routine examination, extensive ultrasonography with color flow profile may be necessary to detect vaginal pseudoaneurysm.Written informed consent was obtained from the patient for publication of this case report and any accompanying images.


Acute lymphocytic leukemia (ALL) is the most common pediatric cancer. The exact cause is not known in most cases, but past epidemiological research has suggested a number of potential risk factors. This study evaluated associations between environmental and parental factors and the risk for ALL in Egyptian children to gain insight into risk factors in this developing country. Methods: We conducted a case-control design from May 2009 to February 2012. Cases were recruited from Children’s Cancer Hospital, Egypt (CCHE). Healthy controls were randomly selected from the general population to frequency-match the cumulative group of cases by sex, age groups  and region of residence , and Upper Egypt (South)). Mothers provided answers to an administered questionnaire about their environmental exposures and health history including those of the father. Odds ratios (ORs) and 95 % confidence intervals (CI) were calculated using logistic regression with adjustment for covariates.adj = 2.5, 95 % CI =1.2 –5.1) and to a lesser extend with her age . Delivering the child by Cesarean section, was also associated with increased risk .Two hundred ninety nine ALL cases and 351 population-based controls frequency-matched for age group, gender and location were recruited. The risk of ALL was increased with the mother’s use of medications for ovulation induction (ORIn Egypt, the risk for childhood ALL appears to be associated with older maternal age, and certain maternal reproductive factors. Worldwide, leukemia is consistently the most common type of childhood cancer in most countries –3, with Regarding pesticides exposure, two meta-analysis have found elevated risk of childhood leukemia with pesticides exposure during pregnancy , 24 . A The aim of the present study was to explore risk factors for childhood precursor B-cell ALL in an Egyptian population by examining potential environmental factors and parental health and reproductive factors that may contribute to the risk of this malignancy.We conducted a case-control study from May 2009 to February 2012. We contacted the parents of 317 eligible cases with a diagnosis of ALL, of whom 299 agreed to participate (94.3 %). Of the 371 population controls approached, the parents of 351 agreed to participate (94.6 %). The recruitment site for cases was the Children’s Cancer Hospital, Egypt (CCHE), which is a referral center for childhood malignancies. The CCHE serves the Cairo metropolitan and surrounding rural and semi-urban areas. Cases were eligible if they resided in Egypt, were diagnosed in the past 6 months with precursor B-cell (PBC) subtype immunophenotype of ALL, and were ≤14 years of age. Healthy controls were randomly selected from the general population to frequency-match the cumulative group of cases by sex, age groups  and region of residence , and Upper Egypt (South)). From these regions, the number of required controls and their characteristics (age and sex) was determined by the sampling frame of the age and sex distributions of the cases; the sampling frame was updated every 6 months as cases were enrolled. The study recruiters visited a randomly selected district  within each region, and approached occupants of houses on a randomly chosen street. If none of the resident's children matched the required sex and age-group, the recruitment team moved to the next house. The recruitment team conducted the controls recruitment during weekends to make sure that the child and mother were at home. Trained interviewers approached the mothers of children, explained the purpose and procedures of the study, and obtained signed consents and assents from the mothers and any child ≥7 years of age, respectively. Interviewers administered the questionnaire face-to-face to mothers of cases and controls. The interview consisted of questions pertaining to the child, the mother and the father, asking about their socio-demographic characteristics, birth and medical conditions of the child, reproductive and medical history of the mothers . We also queried the occupations of the parents and environmental exposures of the parents such as smoking and pesticides, and use of hair dyes. While we were able to collect complete or near-complete data on most variables, birth weight data reported by the mothers were available only for 149 cases and 150 controls .t-test and Chi-squared test to compare continuous and categorical variables, respectively. For the latter tests, Fisher’s exact test was used when any expected cell count was <5. Using case-control status as the outcome variable, we used logistic regression to estimate the adjusted odds ratios (OR) and 95 % confidence intervals (95 % CI) of the associations between ALL and independent variables with adjustment for potential confounders as described below. All statistical tests were performed in Stata v.11 . All P values were two-tailed.We used Student's The distributions of socio-demographic characteristics of the study sample are presented in Table P = 0.08); however, when it was divided into categories , a linear increase in the unadjusted risk with increasing age became evident  for age 23–29 years, and OR = 1.65 (95 % CI = 1.06–2.55) for age 30 year or above, P for trend: 0.02)  for mothers >30 years old). The proportion of mothers with secondary or higher education was higher in cases than in controls ; the association between high educational level and ALL risk was statistically significant  . Treatment for induction of ovulation also showed an association that remained significant after adjustment for age, gender and education (Table We observed differences between cases and controls in Cesarean section and non-term birth Table . None weon Table .P for trend = 0.02). Likewise, the number of siblings also showed an inverse correlation with risk. However, adjustment for maternal education level attenuated the associations of number of deliveries and the number of siblings, and they became non-significant.The number of deliveries, a categorical variable  was associated with decreased risk in childhood ALL; compared to one delivery as the reference, the ORs were 0.83, 0.63 and 0.58 for 2, 3 and 4 + deliveries, respectively  but was not statistically significant after adjustment for maternal educational level.n = 299) with birth weight data, the mean birth weight (cases and controls combined) was 3190.6 g (SEM = 48.8 g (SD: 809.6 g); 95 % CI = 3120.4 – 3312.3). The distribution was not normal and therefore we used non- parametric methods of analysis. The median birth weight was 3000 g (interquartile range: 2750 – 3500 g), which was also the mode. In case-control comparisons, there was a marginally significant difference in birth weight medians between cases  and controls  (P (median test) = 0.06). We did not collect data on gestational age, but we did ask whether the child was born on the expected date or not. Restricting the analysis to those born at expected date (term) yielded a P value of 0.01 for the association between ALL and birth weight. The association was not confounded by any of the variables from the questionnaire. The examination of a potential sex effect on the birth weight association revealed male-specificity: more male cases than male controls had birth weight above the median value , but not in females (47.5 vs 48.5 %). The difference in birth weight association between males and females also reached statistical significance (P for interaction with sex = 0.04).Birth weight data were available for 149 cases and 150 controls . To rule out any selection bias, we compared the cases and controls in this subset for sociodemographic characteristics, and found no difference between the subsamples with and without birth weight data. To validate the representativeness of this subset, we re-examined the associations described above in this subset. All but two associations  remained statistically significant, and all ORs remained in the same direction as in the original analysis (data not shown). In the subjects . Future studies should test hypotheses generated in the present study using a prospective design to address the issues of recall bias and recall errors.To our knowledge, the present study is the most comprehensive and largest study of environmental and maternal risk factors of childhood ALL in a developing country. Studies in such countries, which harbor 90 % of the world's children, are desirable not only to identify unique risk factors, but also to shed light on certain environmental risk factors that may no longer be common enough to study in developed countries. Such clues will help to better identify biological mechanisms in ALL development and pave the way to prevention. Our finding regarding ETS should be considered cautiously since it might represent recall bias until validated in another study.


Transgenerational effects of different environmental exposures are of major interest, with rodent experiments focusing on epigenetic mechanisms. Previously, we have shown that if the study mother is a non-smoker, there is increased mean birth weight, length and body mass index (BMI) in her sons if she herself had been exposed prenatally to her mother's smoking. The aim of this study was to determine whether the prenatal smoke exposure of either parent influenced the growth of the fetus of a smoking woman, and whether any effects were dependent on the fetal sex.Population-based prebirth cohort study.Avon Longitudinal Study of Parents and Children.Participants were residents of a geographic area with expected date of delivery between April 1991 and December 1992. Among pregnancies of mothers who smoked during pregnancy, data were available concerning maternal and paternal prenatal exposures to their own mother smoking for 3502 and 2354, respectively.Birth weight, length, BMI and head circumference.After controlling for confounders, there were no associations with birth weight, length or BMI. There was a strong adjusted association of birth head circumference among boys whose fathers had been exposed prenatally . There was no such association with girls (interaction p=0.006). Similar associations were found when primiparae and multiparae were analysed separately. In order to determine whether this was reflected in child development, we examined the relationships with IQ; we found that the boys born to exposed fathers had lower IQ scores on average, and that this was particularly due to the verbal component .Head size differences concerning paternal fetal exposure to smoking were unexpected and, as such, should be regarded as hypothesis generating. This study is the first to examine the sex-specific fetal effects of parental prenatal exposure to cigarette smoking when the mother herself smoked during pregnancy.Data were collected on a population sample that completed questionnaires blind to the study hypotheses.Birth measurements were undertaken using trained staff with repeated validation.A variety of sensitivity analyses were undertaken, including separate analyses of primiparae and multiparae, as well as of follow-up of the offspring to determine whether the decrement in birth head circumference was reflected in childhood measurement of IQ. All were in accord with the initial finding.The limitation of the study is the failure to obtain comparable data to confirm or negate the study findings.Fetal programming via the mother's nutrition and other aspects of her environment is well recognised as a contributor to adult morbidity and mortalityIn an earlier study of non-smoking mothers, we found an increase in the birth weight and birth body mass index (BMI) of her sons if she had been exposed in utero to her own mothers’ smoking, but there was no such effect if the study father had been exposed in utero.22Our earlier transgenerational study of intrauterine exposure of non-smoking mothers did not consider relationships with fetal growth if the study mother was also a smoker.The current study was therefore carried out to assess whether there is indeed a reduction in the growth of the fetus of a smoking mother if her own mother smoked, and/or whether exposure of the father in utero has any effect on the growth of the child of the smoking mother. In line with the evidence of the accumulating transgenerational human and animal data outlined above, we hypothesise that any effects will differ between boy and girl infants.The data used in these analyses were collected as part of the ALSPAC, which was designed to assess the ways in which the environment interacts with the genotype to influence health and development.25Information collected from the study parents during their study pregnancy included details of the maternal and paternal grandparents. The women and their partners were sent a number of questionnaires during pregnancy.Other data used in the analyses include the study mother's parity ; gestation ; mother's partner smoking at the start of pregnancy ; maternal age at the birth of the child (continuous); housing tenure as a measure of socioeconomic background , maternal education  and maternal alcohol consumption when the mother first felt the baby move .2 (g/m2). In this study, we have used BMI, rather than ponderal index (PI) as our measure of adiposity at birth; although it is traditional to use PI at birth, there is little literature to justify this. It has been suggested that the criteria used to choose whether to use PI or BMI should be a measure that is independent of length.8At delivery, the baby was weighed to the nearest gram; ALSPAC study staff visited the two main delivery hospitals each day and measured the crown-heel length and head circumference of available infants in a standardised manner.Pilot studies before the start of the ALSPAC study had demonstrated that it was usually the student midwife who was given the task of measuring the circumference of the baby's head; she tended to have had little or no training and the measurements made were grossly inaccurate. For this study, we only used measurements that were made by our own staff, after detailed training and with repeated validation over time.Multivariable linear regression models assessed the study children's adjusted mean birth weight, crown-heel length, head circumference and BMI by parental prenatal smoking exposure. All models were adjusted for parity, maternal education, amount mother smoked, paternal smoking in pregnancy and gestation with MGM−M+ and PGM−M+ as the reference categories. Additional models adjusted for maternal age, housing tenure and maternal alcohol use as well as maternal birth weight.In order to determine consistency of the findings, separate analyses were undertaken for primiparae and multiparae. In order to determine whether the head circumference results were biologically meaningful, we also used the fact that reduced head circumference is associated with lower levels of childhood IQ.30In all, there were 3502 births to smoking mothers for whom data were available as to whether their own mothers had smoked when expecting them . ApproxiComparison of data concerning the potential confounders  indicateThe analyses were repeated for primiparae and multiparae separately (see online supplementary tables S1 and S2). The only significant association that remained after adjustment concerned the head circumference of the study sons. The effect sizes were similar for each parity group: for primiparae the effect size was −0.34 cm (95% CI −0.66 to −0.02), p=0.036; for multiparae the adjusted effect size was similar at −0.35 cm (95% CI −0.64 to −0.06), p=0.017. Again there were significant interactions with the sex of the child.Since this association with head circumference was consistent but unexpected, and since there is evidence that birth head circumference is associated with childhood IQ,We investigated whether the parents’ exposure in utero to their own mothers’ smoking was associated with differences in fetal growth among women who smoked in pregnancy, and showed an association between paternal in utero exposure and a reduced head circumference in his sons, but not in his daughters. This was an unexpected finding. A series of sensitivity analyses showed the effect to be almost identical in children born to primiparae and to those born to multiparae. We assessed whether there was confirmatory evidence of an impact on brain size by looking at the IQ of the children. We found a significant reduction in total IQ in 8-year-old boys (but not girls) whose PGM smoked during the pregnancy resulting in the study child's father. The IQ effect size was similar in both parity groups and was still present when birth head circumference was taken into account (data not shown). We showed a stronger association with verbal IQ than performance IQ. To the best of our knowledge, there have been no previous studies that have considered any effects of paternal exposure to smoking in utero on his offspring.There are a number of limitations to this study: (1) details of smoking of parents and grandparents depend on parental self-report—however, there is considerable information to indicate that adults are unlikely to lie about smoking habits, especially when using anonymised self-completion questionnairesincrease in head circumference among the boys born to fathers who had been exposed in utero .8Our parallel study of non-smoking mothers looked at the effect of parental exposure in utero; we found the sons were larger at birth (both in regard to birth weight and birth BMI) if the MGM had smoked in the pregnancy that resulted in the study mother. There was no discernible effect of paternal prenatal exposure on the study child's birth weight or BMI; however, there was a slight IGF2) which is expressed only from the paternally derived chromosome 11, the maternal copy being epigenetically silenced. IGF2 encodes an endocrine and autocrine/paracrine-acting factor important in directing growth during prenatal development.IGF2 DMR  in the newborn infant,IGF2 DMR had been epigenetically modified  by his mother's smoking throughout pregnancy. Furthermore, it is plausible that this epigenetic state could be transmitted via his sperm to the study offspring. Imprinted gene regions tend to escape the usual widespread erasure of DNA methylation from the paternally derived genome in the preimplantation embryo soon after fertilisation.IGF2 DMR in umbilical cord blood being associated with paternal obesity suggesting a preconceptional impact of the obesity (and/or exposures related to it) on the reprogramming of imprint marks during spermatogenesis.37Attributing the smaller head circumference in boys of smoking mothers to the prenatal exposure of the father through his own mother's smoking raises the question of possible mechanisms. How might the information be transmitted via his sperm or in some other way? As we noted in the introduction to this paper, there is increasing evidence that exposures, especially in early life, can lead to enduring changes in the epigenome that, in turn, can modify gene expression. While transgenerational epigenetic inheritance remains controversial, at least in humans,When the mother is a smoker, we found no effect of her own tobacco exposure in utero on the fetal growth of her children. However, when the mother is a smoker, paternal exposure in utero is associated with a reduced head circumference at birth and IQ at 8 years in sons, but not in daughters. We had no prior hypothesis that head circumference would be associated, particularly among sons, so these results must be considered as hypothesis generating, and require testing in further longitudinal data sets.


Novel biomarkers for monitoring progression in neurodegenerative conditions are needed. Measurement of microstructural changes in white matter (WM) using diffusion tensor imaging (DTI) may be a useful outcome measure. Here we report trajectories of WM change using serial DTI in a cohort with behavioral variant frontotemporal dementia (bvFTD).Twenty-three patients with bvFTD (12 having genetic mutations), and 18 age-matched control participants were assessed using DTI and neuropsychological batteries at baseline and ∼1.3 years later. Baseline and follow-up DTI scans were registered using a groupwise approach. Annualized rates of change for DTI metrics, neuropsychological measures, and whole brain volume were calculated. DTI metric performances were compared, and sample sizes for potential clinical trials were calculated.p < 0.001; MD: 2.9%/yr, p = 0.01). MAPT carriers had the greatest change within left uncinate fasciculus ; sporadic bvFTD and C9ORF72 carriers had the greatest change within right paracallosal cingulum . Sample size estimates using FA change were substantially lower than neuropsychological or whole brain measures of change.In the bvFTD group as a whole, rates of change in fractional anisotropy (FA) and mean diffusivity (MD) within the right paracallosal cingulum were greatest (FA: −6.8%/yr, Serial DTI scans may be useful for measuring disease progression in bvFTD, with particular trajectories of WM damage emerging. Sample size calculations suggest that longitudinal DTI may be a useful biomarker in future clinical trials. The behavioral variant of frontotemporal dementia (bvFTD) is a common cause of early onset dementia with progressive decline in personality and interpersonal skills, with gradual emergence of abnormal behaviors including apathy, obsessionality, and loss of empathy.2To date, longitudinal magnetic resonance imaging (MRI) has been shown to be a useful biomarker in neurodegenerative diseases, given its wide availability, ease of interpretation, and sensitivity in detecting change  over time. Previous longitudinal imaging studies of bvFTD have used structural MRI to measure rates of whole brain and ventricular change.7A further problem in tracking progression in bvFTD is its broad pathological and clinical heterogeneity. Predicting underlying pathology on clinical or radiological grounds remains challenging. As such, the use of current clinical or neuroimaging measures to evaluate treatments, which will likely target specific molecular pathologies, is problematic. A number of recent reports have proposed a common "network-led" framework to understand how neurodegenerative diseases evolve.This study aims to address these issues by investigating longitudinal white matter change in key white matter structures in a group of patients with bvFTD. We also investigate the utility of DTI as a potential biomarker for clinical trials by comparing DTI measures of change with established MRI and neuropsychological measures and estimate sample size requirements for potential future clinical trials.(MAPT), progranulin (PGRN), and chromosome 9 open-reading frame 72 (C9ORF72) genes. Twenty-three participants were identified as fulfilling inclusion criteria. Four participants were not included in the DTI analysis for the following reasons: 2 had significant artifact on follow-up DTI scans, 1 had an incomplete sequence due to scanner intolerance, and another had a sphenoid wing meningioma. Eighteen cognitively normal age- and gender-matched participants, with no history of psychopathy, stroke, myocardial infarct, or peripheral vascular disease, underwent the same test batteries as those with bvFTD. Each study participant underwent a battery of neuropsychological tests at baseline and follow-up, including assessments of general intellectual functioning, verbal and visual memory, naming, visuospatial and visual perception, calculation, and executive function. Social cognition was assessed using abbreviated versions of the emotion recognition (first 14 items) and social inference (first 9 items) subsets of the Awareness of Social Inference Test.11Patients were recruited from 2009 to 2014 as part of a prospective study tracking disease progression in patients suspected to have frontotemporal lobar degeneration at the Specialist Cognitive Disorders Clinic, National Hospital for Neurology and Neurosurgery, London, United Kingdom. Patients who met current consensus criteria2). Field maps, b = 0 s/mm2 images, and a sagittal 3-dimensional magnetization prepared rapid gradient echo T1-weighted volumetric MRI  were also acquired. Following visual inspection for artifacts, field map–based unwarping was applied to the diffusion-weighted images and they were affine-aligned to the average b0 image using FLIRT within the FMRIB Software Library (FSL v5.0.1 reference FSL) to correct for motion and eddy currents.13Serial MRI scans were performed on the same Siemens Trio  3T MRI scanner using a 32-channel phased array head coil. Two 64-direction DTI sequences were acquired and averaged using a single-shot, spin-echo echo planar imaging sequence  software package,Registration of DTI images was carried out using a previously published method carried out on the same magnetic resonance scanner  cingulum bundle, paracallosal  cingulum bundle, corticospinal tract, superior cerebellar peduncle, and fornix, were located using the white matter labels. Binary masks of each region were generated using a threshold of 50% on the white matter probability map with a further 1mm erosion around the boundary of each mask to provide high anatomical specificity. A visual inspection of each mask was performed to ensure appropriate coverage. The uncinate fasciculus did not undergo erosion, as this would have limited the mask size, reducing sensitivity to detect change in this tract.Volumetric image analysis was performed using a rapid, semiautomated segmentation technique yielding a brain region separated from surrounding cerebrospinal fluid, skull, and dura. Serial scans were aligned and volume change calculated directly using the boundary shift integral (BSI).Statistical analyses were carried out using Stata 12 . Cross-sectional DTI metric data were compared between disease and cognitively normal groups using a linear regression model adjusting for age, gender, and disease duration. Mixed-effects linear regression models with random intercept were used to compare longitudinal change between groups for each DTI metric and region of interest, adjusting for age, gender, and disease duration. For longitudinal models log of each DTI metric was the dependent variable, with disease group, time from baseline scan in years, and interaction between disease group and time included to provide estimates of differences in the rate of change as a percentage per year. This methodology was also used to compare cross-sectional and longitudinal neuropsychological data between groups.To determine the accuracy, sensitivity, and specificity of each DTI metric in classifying individual participants into separate groups (bvFTD or control), receiver operating characteristic curves were constructed using either raw DTI metric data (for baseline measurement) or the estimated mean difference in the rate of change for each diffusivity metric . Areas under the curve (AUCs) were calculated for regions of interest, which were significantly different when compared with controls.Estimation of sample sizes for future trials were calculated (using Stata) with 80% power and 5% 2-tailed significance using the mean difference in the rate of change between groups for each DTI metric and neuropsychological score, and whole brain atrophy, using the BBSI, to detect a 20, 30, 40, and 50% reduction in yearly change.MAPT ; 4 patients had a C9ORF72 mutation. Patients and cognitively normal participants were matched for age, gender, and total intracranial volume. Compared with cognitively normal participants, those with bvFTD had significantly lower Mini-Mental State Examination scores (p < 0.01) and whole brain volumes (p ≤ 0.001) at baseline and follow-up. Rates of atrophy were greatest (p = 0.002) in the bvFTD group, with the highest rate of volume loss occurring in those with MAPT mutations (p = 0.001).Demographic and volumetric imaging characteristics of study participants are shown in Tablep < 0.001).Neuropsychological performance at baseline and longitudinally is shown in Tablep ≤ 0.02) FA and increased MD within the body of the corpus callosum, bilateral uncinate fasciculus, and right parahippocampal cingulum, and additionally increased MD in left parahippocampal cingulum. In MAPT mutation carriers, FA was 6.4%  lower in right parahippocampal cingulum compared with controls, whereas MD was 17.7% higher in right uncinate fasciculus  and 9.4% higher in right parahippocampal cingulum . In sporadic bvFTD, FA was 10.3% lower in left uncinate fasciculus  and 6.7% lower in right parahippocampal cingulum  compared with controls, whereas MD was 31% higher in left uncinate fasciculus  and 15% higher in right parahippocampal cingulum . In C9ORF72 mutation carriers, FA was 11% lower in both right and left superior cerebellar peduncle  compared with controls.Cross-sectional DTI metric data for FA and MD are shown in Tablep < 0.001; left, −5.5%/yr, 95% CI = −6.9 to −2.2%, p < 0.001) and bilateral uncinate fasciculus . The largest increases in MD were within bilateral uncinate fasciculus  and bilateral parahippocampal cingulum .Rates of change for each region of interest and diffusivity metric are shown in Tablep < 0.001; left: −7.9%/yr, 95% CI = −12.0 to −3.7%, p < 0.001) in MAPT mutation carriers; bilateral paracallosal cingulum bundle  in those with sporadic bvFTD, and in right paracallosal cingulum bundle  in C9ORF72 mutation carriers. Compared with cognitively normal participants, the largest increases in MD were in bilateral uncinate fasciculus  in MAPT mutation carriers, bilateral paracallosal cingulum bundle  in sporadic bvFTD, and left uncinate fasciculus  in C9ORF72 mutation carriers.Rates of change for each region of interest and diffusivity metric are shown in TableAUC data indicated that classification of control and bvFTD groups cross-sectionally were best achieved using RD, measured within the right uncinate fasciculus , followed by MD , FA , and AX . Classification of control and bvFTD groups longitudinally were best achieved using FA change, measured within the right cingulum bundle , followed by MD , and RD , with AX performing less favorably .Sample sizes required for future clinical trials were calculated using annualized change score in 3 potential outcome measures: whole brain atrophy (using BBSI), change in graded naming test, and DTI change within either right paracallosal cingulum or right uncinate fasciculus . Sample size estimates corrected for control rates of change are displayed in TableThis study is among the first to demonstrate the feasibility of carrying out longitudinal DTI in patients with bvFTD. Compared with previous DTI studies, we used what we believe to be an improved approach for DTI spatial normalization, which enforces longitudinally and cross-sectionally consistent and accurate region of interest segmentations, thus reducing potential noise within the DTI data set, which is often a significant methodological problem. Using these improved DTI methods, we (1) report both core and mutation-specific patterns of white matter change in bvFTD; and (2) demonstrate that longitudinal DTI is a feasible outcome measure for clinical trials, requiring smaller sample sizes than other more conventional outcome measures.We found that decreasing FA and increasing MD within bilateral paracallosal cingulum bundle, body of the corpus callosum, and bilateral uncinate fasciculus was the most consistent finding across all bvFTD groups. These findings are in line with a number of cross-sectional DTI studies in bvFTD.MAPT mutation carriers, the most robust changes were within bilateral uncinate fasciculus, bilateral paracallosal and parahippocampal cingulum, and the body of the corpus callosum; in the C9ORF72 mutation carriers, within corpus callosum and right paracallosal cingulum bundle; and in sporadic bvFTD, within the bilateral paracallosal cingulum bundle. MAPT mutation carriers were the only group to have significant change over time within both right and left uncinate fasciculus, with a high burden of white matter change occurring in medial temporal lobe regions. Breakdown of these tracts is biologically plausible, given that these tracts link gray matter structures that show preferentially more atrophy in affected MAPT mutation carriers.MAPT mutation carriers.MAPT mutation carriers.C9ORF72 group must be viewed in terms of the small sample size, it is of interest that only this group showed significant difference in bilateral superior cerebellar peduncles cross-sectionally and neared significance (p = 0.06) longitudinally. This might be surprising for a “frontotemporal” dementia; however, this finding is consistent with other data suggesting that changes within the cerebellum and associated structures may be a particular hallmark of C9ORF72 mutation carriers.Different patterns of white matter change were associated with different FTD subtypes. In An important issue that remains poorly understood is the choice of DTI metric to detect and track white matter change. To date, most longitudinal DTI studies have reported only changes in FA,Understanding how disease neurobiology results in specific changes to DTI metrics will require further studies with larger pathologically confirmed cohorts, including the progranulin genetic subtype, which the current study lacks. Larger studies will be required to confirm these data as well as to examine changes within other white matter structures not included here. Further improvements to image acquisition and analysis through the use of multishell acquisitions may improve spatial resolution of the data and provide greater information on white matter integrity.39We have demonstrated that a within-subject measure of DTI change is a potentially useful disease biomarker with the ability to detect greater differences across white matter regions compared with cross-sectional measures. This, coupled with lower sample size requirements, suggests that longitudinal DTI may be an important biomarker for disease monitoring with particular implications for future clinical trials.


Smoke-free legislation is associated with improved early-life outcomes; however its impact on perinatal survival is unclear. We linked individual-level data with death certificates for all registered singletons births in England (1995–2011). We used interrupted time series logistic regression analysis to study changes in key adverse perinatal events following the July 2007 national, comprehensive smoke-free legislation. We studied 52,163 stillbirths and 10,238,950 live-births. Smoke-free legislation was associated with an immediate 7.8%  reduction in stillbirth, a 3.9%  reduction in low birth weight, and a 7.6%  reduction in neonatal mortality. No significant impact on SIDS was observed. Using a counterfactual scenario, we estimated that in the first four years following smoke-free legislation, 991 stillbirths, 5,470 cases of low birth weight, and 430 neonatal deaths were prevented. In conclusion, smoke-free legislation in England was associated with clinically important reductions in severe adverse perinatal outcomes. Fifty years after the official recognition of its adverse health effects, tobacco smoking remains the primary cause of preventable mortality worldwide1141Population-level reductions in SHS exposure and active smoking can effectively be achieved by implementing legislation to prohibit smoking in enclosed public places and the workplace7910111213We sought to address this by investigating the association between the 2007 national introduction of smoke-free legislation in England and various indicators of early-life mortality – namely stillbirth, neonatal mortality, and SIDS – in a national birth cohort of over 10 million singleton births. Given the conflicting evidence from previous studiesDuring the study period, 10,291,113 singleton births were registered in England, including 52,163 stillbirths 0.5%; . Among tStillbirth and early-life mortality were highly associated with low birth weight categories and low socioeconomic status, and less strongly with other demographic characteristics . ImportaIntroduction of smoke-free legislation was associated with an immediate 7.8%  reduction in odds of being stillborn, a 3.9%  reduction in odds of having low birth weight, and a 7.6%  decrease in odds of neonatal mortality over-and-above the underlying temporal trend . No signThe reduction in neonatal mortality was primarily attributable to an impact on late neonatal deaths: −13.7%  may have resulted in unexplained variation43Whereas we identified an overall association between smoke-free legislation and reduced infant mortality, we were surprised by the finding that SIDS was not affected1The impact of smoke-free legislation on perinatal and early-life health is likely to have been mediated via several routes. A number of previous studies have demonstrated important drops in maternal smoking during pregnancy following implementation of smoke-free legislation110111213474849Our findings add an important new dimension to the emerging evidence on the benefits of smoke-free legislation to child health971821Since the majority of the burden of early-life morbidity and mortality occurs in low- and middle-income countries18In conclusion, we present evidence that implementation of smoke-free legislation in England was associated with substantial perinatal and early-life benefits, with over 5,000 cases of low birth weight and almost 1,500 deaths averted within four years.This study was performed according to a pre-specified and registered research protocol . Using a national dataset of all singleton births registered in England between 1995 and 2011, we analysed the association between the July 2007 implementation of smoke-free legislation and the odds of stillbirth, low birth weight, and early-life mortality.This study was reviewed by the National Health Services (NHS) South East Scotland Research Ethics Service and The University of Edinburgh’s Centre for Population Health Sciences Ethics Review Group. Both committees provided an exemption from formal ethical assessment based on the use of anonymised, unidentifiable data.The intervention under study was the implementation of smoke-free legislation in England on 1 July 2007Four primary outcomes were evaluated: low birth weight ; stillbirth (intrauterine death from 24 weeks of gestation); neonatal death (death in the first 28 days of life); SIDS  10-U R95, or R99 with no other specification).In addition, we considered the following secondary outcomes: very low birth weight ; early neonatal death (death in the first week of life); late neonatal death (death between 7 and 28 days of life); post-neonatal death (death between 28 days of life and the first birthday); and infant death (death within the first year of life).a priori not to involve preterm birth as an outcome or intermediate variable in our analyses.Data on gestational age were not available through the Office for National Statistics before 2006, restricting the pre-legislation period to only 1.5 years. Given the recognised seasonal variation in gestational age we considered this period insufficient to estimate underlying trends pre-legislation, and therefore decided Individual-level data on all registered singleton births in England between 1 January 1995 and 31 December 2011 were obtained via the government’s Office for National Statistics. Data were linked by the Office for National Statistics to death certificates for stillbirths and for all infants dying before their first birthday. To minimise the risks of individuals being identified, individual-level continuous variables were categorized as follows: month and year of birth; month and year of death; timing of death ; sex; birth weight ; maternal marital status (married /not married); maternal parity; and maternal age (<20 years/20–24 years/25–29 years/30–34 years/35–39 years/≥40 years). The individual-level data also included Government Office Region, urbanisation level , and socioeconomic status measured using the Index of Multiple Deprivation (in quintiles), these being derived from maternal post code of residence. The Index of Multiple Deprivation is a relative measure of deprivation at the small-area levelThe patterning of demographic data was initially tabulated by timing of death and birth weight categories. Separate logistic regression models using individual-level data were then developed to investigate the association between the introduction of smoke-free legislation and the odds of developing each outcome5151The following categorical covariates were included in each model to account for potential individual-level confounding: sex; maternal age; maternal marital status; Index of Multiple Deprivation quintile; region; and urbanisation level. All mortality-related outcomes were also adjusted for birth weight.Among the model covariates, data on parity were missing for a substantial portion of the study population (41.3%), as this variable had been recorded for married women only. All other covariates had 0–3% missing data. Since logistic regression analysis needs to be undertaken on cases with complete data for all covariates, parity was excluded from the primary models to maximise power. We performed two sets of pre-specified sensitivity analyses to investigate potential residual confounding resulting from exclusion of parity as a covariate. In the first set, models were re-run with parity included, analysing individuals with complete data only. In the second set, multiple imputations were performed using chained equations to impute missing data for parity, creating five unique datasetsA change in odds of developing low birth weight following the introduction of smoke-free legislation is potentially on the causal pathway between smoke-free legislation and any observed impact on early-life mortality . We therIn order to estimate the absolute impact of smoke-free legislation for outcomes that were significantly affected, we developed counterfactual scenarios51All analyses were undertaken using Stata SE version 12.0 .As we used the maximum time span and population available, sample size calculations were redundant. Given the national nature of this evaluation, we estimated that we would have adequate power to detect clinically relevant temporal changes in the outcomes of interest11121323242527How to cite this article: Been, J. V. et al. Impact of smoke-free legislation on perinatal and infant mortality: a national quasi-experimental study. Sci. Rep.5, 13020; doi: 10.1038/srep13020 (2015).


Some studies have noted an association between maternal occupational exposures to chlorinated solvents and birth defects in offspring, but data are lacking on the potential impact of industrial air emissions of these solvents on birth defects.With data from the Texas Birth Defects Registry for births occurring in 1996–2008, we examined the relation between maternal residential proximity to industrial air releases of chlorinated solvents and birth defects in offspring of 60,613 case-mothers and 244,927 control-mothers. Maternal residential exposures to solvent emissions were estimated with metrics that took into account residential distances to industrial sources and annual amounts of chemicals released. Logistic regression was used to generate odds ratios and 95% confidence intervals for the associations between residential proximity to emissions of 14 chlorinated solvents and selected birth defects, including neural tube, oral cleft, limb deficiency, and congenital heart defects. All risk estimates were adjusted for year of delivery and maternal age, education, race/ethnicity, and public health region of residence.Relative to exposure risk values of 0, neural tube defects were associated with maternal residential exposures  to several types of chlorinated solvents, most notably carbon tetrachloride ; chloroform ; ethyl chloride ; 1,1,2-trichloroethane ; and 1,2,3-trichloropropane . Significant associations were also noted between a few chlorinated solvents and oral cleft, limb deficiency, and congenital heart defects. We observed stronger associations between some emissions and neural tube, oral cleft, and heart defects in offspring of mothers 35 years or older, such as spina bifida with carbon tetrachloride , cleft palate with 1,2-dichloroethane , cleft lip with or without cleft palate with ethyl chloride , and obstructive heart defects with trichloroethylene .These findings suggest that maternal residential proximity to industrial emissions of chlorinated solvents might be associated with selected birth defects in offspring, especially among older mothers. Oxidative stress has been suggested as a mechanism by which some teratogens cause birth defects, and sevSeveral epidemiologic studies have examined the relation between maternal occupational exposure to various chlorinated solvents and birth defects in offspring–14 with In the present study, we examined whether maternal residential proximity to industrial air emissions of chlorinated solvents was associated with neural tube, oral cleft, limb deficiency, and congenital heart defects in offspring. We accounted for residential proximity to multiple facilities and the annual amounts of specific chlorinated solvents released from each. We also examined whether older maternal age modified associations between residential proximity to chlorinated solvent emissions and birth defects with one of the largest study populations to date.Live births, fetal deaths, and induced terminations with selected congenital malformations, including neural tube defects, oral cleft defects, limb deficiencies, and congenital heart defects, were identified from the Texas Birth Defects Registry (TBDR) for births occurring during 1996–2008. The Registry used active surveillance in which trained staff reviewed medical records from hospital, birth centers, and midwifery locations. TBDR covered all cases that were live births, spontaneous fetal deaths (reportable in Texas if 350 grams or more or length of gestation 20 weeks or greater), and pregnancy terminations. Cases were selected in the study if they were delivered between 1996 through 2008, had a maternal address in Texas at time of delivery, and had a diagnosis of one or more of the selected birth defects. The majority of case deliveries were live births (97.1%) with 1.4% classified as spontaneous fetal deaths, 1.5% induced terminations of pregnancy, and 0.1% unspecified fetal deaths. Live birth and fetal death cases were linked to their respective birth or fetal death certificates that contained information about maternal residential address at the time of delivery and demographic characteristics.Birth defects were further classified as either isolated or not isolated. Isolated defects were defined as those having a single major defect alone or together with one or more of a list of minor defects used in the National Birth Defects Prevention Study.Control-births, defined as live births without any recorded birth defects, were randomly selected from live birth certificate data obtained from the Center for Health Statistics at the Texas Department of State Health Services. These births were frequency matched to cases by year of delivery (1996–2008) and public health service region (11 regions) in which the case-mothers resided at the time of delivery. We frequency matched birth defect cases and controls by public health region because between 1996 and 1999, active surveillance of birth defect cases did not occur statewide, but was confined to certain regions, including 2 regions in 1996, 6 regions in 1997, and 8 regions in 1998. These regions range from 12,060 (Region 5) to 61,456 square miles (Regions 9/10 combined).ESRI ArcGIS 8.3 ArcMap 9.3 was used to geocode the maternal addresses obtained from live birth and fetal death certificates. A total of 86.7% of the maternal addresses for case-mothers and 87.2% for those of control-mothers were successfully geocoded to street level.Data regarding air emissions from Texas industrial facilities were obtained from the U.S. Environmental Protection Agency Toxic Release Inventory (TRI) program. The online TRI databases contain names of facilities required to report under section 313 of the Emergency Planning and Community Right-to-Know Act and include information about location, reporting year, chemicals released, and estimated pounds per year released into various environmental media, such as air and water bodies. For the present project, industrial air releases of 14 chlorinated solvents were identified for study, including carbon tetrachloride; chloroform; 1,1-dichloroethane; 1,2-dichloroethane; 1,2-dichloroethylene; ethyl chloride; methyl chloroform; methylene chloride; perchloroethylene; propylene dichloride; tetrachloroethane; 1,1,2-trichloroethane, trichloroethylene; and 1,2,3-trichloropropane.Based on the addresses available in the USEPA TRI databases, we geocoded the TRI facilities in three steps. The addresses were first geocoded using Centrus Geocoder for ArcGIS. Then, the TRI addresses that were not geocoded in the first step were geocoded by the default geocoding tools from ArcGIS 10.0 using the Census 2000 street map as the reference layer. Finally, the remaining TRI addresses that were not successfully geocoded using the first two methods were geocoded with Google map geocoding API. Only TRI addresses geocoded to the premise level were considered successful matches. Over the 13 year period of 1996–2008, approximately 90% of the industrial facilities, on average, were successfully geocoded.θ at location i from all emission sources is given by Expression (1).To estimate potential intensities of various chlorinated solvent air releases in the vicinity of maternal residences, a modified version of the Emission Weighted Proximity Model (EWPM) was used. This moAθi is the estimated quantity of air pollutant θ at location i from all emission sources j  within the threshold distance. This quantity was used as a proxy to represent the exposure risk value of a person at location i to air pollutant θ; m is the number of emission sources relevant to a person at location i in the area in question; Eθij is the emission rate of air pollutant θ from any emission source j that is within the effective threshold distance related to location i; Tθij is the duration of emission of air pollutant θ from emission source j; kθ is effective threshold distance beyond which air pollutant θ is considered to have no harm to an individual and is determined by the physical and photochemistry characteristics of the pollutant in question. In addition to the limitation of chemical quantities aggregated into pounds per year, further misclassification could have been introduced if the amounts released per year varied since the first trimester of some pregnancies might have been during the preceding year. Furthermore, length of gestation for fetal deaths and induced terminations (cases) would have been shorter than that for the majority of live births, although approximately 97% of the cases were liveborn and the most vulnerable period for teratogenesis is the first trimester.While the Texas Birth Defects Registry includes elective terminations in their active surveillance of birth defects, we were unable to include these cases in this study because of missing data from lack of linkage to a vital record. This lack of linkage had the greatest impact on our analyses of the relation between chlorinated solvent emissions and neural tube defects in which 69% of maternal addresses were geocoded to street level and 22% were missing information regarding maternal education, a variable strongly related to maternal residential proximity to industrial facilities along with maternal race/ethnicity.With respect to specific defects, the highest proportion of induced terminations were found with anencephalic cases (39.7%) with percentages less than 5% for the other studied defects with the exception of spina bifida (8.3%) and lower limb reduction defects (5.6%). Given the high proportion of anencephalic cases that could not be analyzed because of missing information, findings related to this defect should be interpreted cautiously because it is not known whether women who undergo induced terminations are more or less likely to live near sources of industrial emissions.In this study, we were unable to account for recurrence of birth defects in subsequent pregnancies of case-women, a known risk factor for birth defects. RecurreThe large sample size of this study allowed examination of associations between specific chlorinated solvents and birth defects as well as stratification by maternal age. Furthermore, a refined exposure assessment accounted for multiple facilities and their respective annual quantities of chemical releases, an improvement from previous studies that estimated exposure based on distance to the nearest industrial facility.Study findings suggest that maternal residential proximity to air emissions of chlorinated solvents are associated with neural tube, oral cleft, and congenital heart defects, especially among offspring of older mothers. To better understand the relation between maternal exposure to chlorinated solvent emissions and birth defects in offspring, we recommend that future studies focus on populations in which air measurements of these chemicals and residential histories during pregnancy are available.


Rodents are a useful model for life science research. Accumulating evidence suggests that the offspring of mice and rats suffer from similar disorders as humans when exposed to hypoxia during pregnancy. Importantly, with antenatal hypoxic exposure, human neonates demonstrate low birth weight or growth restriction. Similarly, with antenatal hypoxic exposure rodents also demonstrate the fetal growth restriction (FGR). Surprisingly, there is no consensus on the minimum duration or degree of hypoxic exposure required to cause FGR in rodents. Thus, we have reviewed the available literature in an attempt to answer these questions. Based on studies in rats, birth weight reduction of 31% corresponded to 10th percentile reduction in birth weight curve. With the similar criterion (10th percentile), in mice 3 days or more and in rats 7 days or more of 14% or lower hypoxia administration was required to produce statistically significant FGR. Antenatal stressors during pregnancy can lead to fetal diseases as well as the programing of disorders that occur in adulthood Barker, . Hypoxicwww.cdc.gov) low birth weight (LBW) has been defined as a birth weight of less than 2500 g. LBW could be a result of genetic background without any obvious factor inducing it. In such cases, the LBW infant is referred to as small for gestational age (SGA). However, placental, fetal, or maternal factors may be responsible for LBW. In such cases, in which the fetus is unable achieve its full genetic potential as a consequence of some known factor is diagnosed as FGR. FGR is one of the major complications of antenatal hypoxic exposure in humans, and defined as fetal weight below the 10th percentile of the population for its gestational age resulting from a pathological cause. Of note, FGR affects 7–10% of pregnancies or about 20.5 million infants each year  . Both litter size and placental weight were also significantly reduced. Furthermore, in a recent study, 12–14% of oxygen for 7 days  lead to 31% (P < 0.050) reduction in pups birth weight  with 12 and 14% oxygen administration on DPC 18–21 , compared to normoxic control to label the pups as FGR. In more than 100 control litters, a group of investigators further determined how this statistically significant weight translated into percentile, and defined FGR as birth weight below the 15th percentile of sex-matched controls  is the time frame most studied for hypoxic insult, but earlier exposure may also lead to FGR. Importantly, we propose that in rodents, at least 22% or more reduction in pups birth weight should be considered as FGR.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.


In addiIn-vivo CMRE was performed on a pig in a 1.5T MRI scanner. CMRE imaging parameters: TE/TR=9.7/21.4; flip angle 25○; mechanical frequency 80 Hz; encoding frequency 160 Hz. Post CMRE acquisition the heart was arrested in diastole using potassium chloride and in-situ cardiac DTI was performed. Cardiac DTI parameters: TE/TR=80/3200; flip angle 90○; b-value=0/1000 s/mm2; number of directions=12; number of averages=10. CMRE and DTI was performed at the same resolution and the parameters were FOV=320mm3; imaging matrix 128x128; slice thickness=2.5mm; DTI was registered with CMRE to exactly match the voxel information from both sets of acquisition. Then both CMRE and DTI were masked to extract the left ventricle. Masked images were processed to estimate i) principle eigenvectors from DTI data sets; ii) and first harmonic displacements from CMRE wave data. Next, a spatial spectral filter was applied on the first harmonic displacement data to isolate waves traveling in particular directions defined by the principle eigenvector. Simultaneously, Helmholtz decomposition was performed to separate the filtered displacements into its longitudinal and transverse components. An orthotropic inversion [11,C22,C33) and shear  stiffness coefficients.Waveguide CMRE involves performing diffusion tensor imaging (DTI) in conjunction with conventional CMRE. Figure in-vivo anisotropic stiffness using CMRE. However further validation and application in a diseased model is required.We have demonstrated the feasibility of estimating This study has been supported by AHA 13SDG14690027.