Spirodela polyrhiza (Greater Duckweed), provides an opportunity to study diversification because although partially reliable seasonal cues exist, its growing season is subject to an unpredictable and literally “hard” termination when the surface water freezes, and overwinter survival depends on a switch from production of normal daughter fronds to production of dense, sinking “turions” prior to freeze‐over. The problem for S. polyrhiza is that diversified dormancy behavior must be generated among clonally produced, genetically identical offspring. Variation in phenology has been observed in the field, but its sources are unknown. Here, we investigate sources of phenological variation in turion production, and test the hypothesis that diversification in turion phenology is generated within genetic lineages through effects of parental birth order. As expected, phenotypic plasticity to temperature is expressed along a thermal gradient; more interestingly, parental birth order was found to have a significant and strong effect on turion phenology: Turions are produced earlier by late birth‐order parents. These results hold regardless of whether turion phenology is measured as first turion birth order, time to first turion, or turion frequency. This study addresses a question of current interest on potential mechanisms generating diversification, and suggests that consistent phenotypic differences across birth orders generate life history variation.Environmental unpredictability is known to result in the evolution of bet‐hedging traits. Variable dormancy enhances survival through harsh conditions, and is widely cited as a diversification bet‐hedging trait. The floating aquatic plant,  Organisms may respond directly to environmental change through (nonadaptive) physiological or behavioral plasticity . Some organisms rely on environmental cues to trigger the onset and/or termination of dormancy. Cues that can influence dormancy behavior include photoperiod , or Greater Duckweed, is a widely distributed floating perennial plant found in a diverse range of aquatic environments  sequentially through budding from two meristematic pouches, although as many as twenty have been observed in other studies , S. polyrhiza was found to produce only vegetative fronds at water temperatures above 18°C and mainly turions at temperatures less than 10°C. The goal was thus to determine plasticity to temperature for different birth‐order mothers over a range that induces the production of both fronds and turions.A thermogradient incubator, generally used to investigate seed germination in response to temperature,  for at least five generations. Following this standardization, budding of fronds was monitored within the 12‐well plates. Each complete birth‐order sequence  was produced by a unique mother. All five required birth orders were separated and transferred to each of two replicate, color‐coded strainers at each of seven temperatures within the thermogradient resulting in a total of 70 strainers (35 per lane) and a sample size of four fronds per strainer giving 280 fronds per thermogradient run, or a total of 560 fronds. Odd birth orders were used because fronds are produced from two pouches almost simultaneously . The thermogradient remained stable throughout the study, and the growth medium was replaced at the introduction of each new birth order and once per week thereafter. However, to ensure that temporal factors did not confound birth‐order effects, two approaches were taken. First, an “extension run” was performed immediately following the first run without altering the conditions, but using only birth‐order 1 mothers so that their behavior could be compared to the original birth‐order 1 mothers from the beginning of the run. In addition, a second complete thermogradient run was performed, but as a “backward run” in which birth orders were introduced into the thermogradient in reverse sequence . This reverse birth order had to be produced while manipulating the timing of placement into the thermogradient to be similar to that of the forward run. To prepare birth‐order 9 mothers to initiate the backward run, birth orders 1 through 8 were discarded as they were produced in their 12‐well plates. New plates were setup to prepare birth‐order 7 fronds, timed for their correct sequential introduction into the thermogradient, by discarding birth orders 1 through 6. All birth orders were similarly prepared, so that each birth order was transferred to the thermogradient with similar, but reversed, timing compared to the forward run. Thus, unlike in the forward run, daughters were necessarily derived from different mothers, although from mothers of the same clone and with the same birth‐order history. In this run, the medium was replenished once a week and the placement of a new birth order inside the incubator was scheduled so that it would fall on the same day as the medium change. Birth‐order 1 fronds were thus placed into the thermogradient 28 days after birth‐order 9 fronds were first introduced.r = .83), partial correlations with birth order indicate that the negative relationship between mother size and birth order is accounted for a highly significant partial correlation with width .Turion production for the different birth‐order mothers was measured as birth order of first turion produced, as days to first turion, and as turion frequency. Once a mother frond produced a daughter, whether a frond or turion, it was gently separated from the mother only upon maturity, that is, after the daughter initiated its own daughter frond. Length and width of all mother fronds were also measured to assess the effect of size on turion phenology. Each mother and daughter produced was photographed using an Olympus SZX12 microscope with an Infinity 3 Lumenera camera using a 0.5× lens at 10× magnification. The digital photographs were calibrated, and frond length (apex to opposite edge) and width (perpendicular to length at widest point) measurements were taken using NIH ImageJ version 1.49v. Here, we use mother width as our size measure because preliminary analyses show that although length and width are strongly correlated , and their interactions on first turion birth order was determined using ANCOVA. Separate post hoc multiple regression analyses were conducted for the forward and backward runs to further examine the effect of temperature and mother birth order within each run in the event of interactions with thermogradient run. The extension run was used to compare the effects of mother birth order to the forward run. Mother birth order 1 of the extension run was coded as a separate level in an ANCOVA of the effects of mother birth order  and temperature (continuous) on first turion birth order that included all data from the main thermogradient study and the extension run. A post hoc Tukey's test was then used to compare behavior across birth orders.Days to first turion and turion frequency  are alternate measures of turion phenology for individual mothers. For completeness, these were similarly assessed by conducting multiple regression analyses of the effect of temperature, mother birth order, and their interactions for the forward and backward runs. The phenology of turion production (first turion birth order and days to first turion) would be seriously biased toward early turion production if turion data for mothers producing no turions during their lifetime were simply missing from the dataset. To accommodate standard analyses consistently across the study, a conservative  transformation approach was taken to eliminate the “invisible fraction” bias Weis,  caused bFinally, logistic regression was conducted to predict the effect of temperature, mother birth order, and daughter birth order on daughter state being a frond or turion. All analyses were performed using 3p < .0001), and their interaction becomes significant ; it is clear .The ANCOVA to assess the effects of temperature, mother birth order, size, and thermogradient run on first turion birth order found highly significant effects of temperature and mother birth order, and a marginally significant effect of frond size Table . Howeverr Figure  that higr Figure . These rF = 13.52; p < .001) occur between mother birth order 1 of the extension run and all mother birth orders  except birth order 1 of the main thermogradient study in a post hoc Tukey's test.Analysis of covariance—with temperature as the covariate—comparing the “extension” run (using only birth order 1 mothers) and the forward run finds that significant differences in first turion birth orders among mother birth orders  and birth order (both runs: p < .001) on turion frequency, Figure p < .001; Figure p = .277; Figure p = .130; turion frequency: p = .913).Alternate measures of turion phenology and production are the time taken to produce a first turion, and the frequency of turion formation. Experiment‐wise, individual mother fronds produced between zero and five turions , with a substantial proportion producing two (29%) or three 24%) turions, with only 5% producing four or five. Multiple regression analyses to assess the effects of temperature, mother birth order, and their interactions on days to turion production  produced as a function of mother birth order, daughter birth order, and temperature. All effects are strong predictors of the formation of turions Table . Based o4Phenotypic plasticity and bet hedging are two evolutionary modes of response to varying environmental conditions, where adaptive plasticity may evolve in response to cues , by time elapsed since production, by turion birth order, or by other factors, and is the subject of ongoing research.Fourth, this study of the phenology of turion formation addresses mechanisms of coping with the onset of winter, but another component of uncertainty in the Lemna spp. exhibit diminishing fitness with birth order; if this holds for S. polyrhiza, the switch to turion production in late birth‐order fronds with decreasing residual reproductive value may be a low‐cost solution for variance generation. Interest in the adaptive significance of the expression of individual phenotypic variance (Bull, e.g., Rovira‐Graells et al., The duckweed system provides opportunities to further test variance strategies, as well as life history theory more broadly. For example, Barks and Laird  demonstrce Bull,  has recece Bull, , and hasBoth authors contributed equally to this work. H.S.M. and A.M.S. designed the study, developed the methodology, performed the analysis and wrote the manuscript. H.S.M. performed the experiments.


It is discovered that the exchange interaction can be substantially weakened under the influence of electric field applied perpendicular to the magnetic film surface. Furthermore, we demonstrate that the electric field modified exchange constant could effectively control the propagation of spin waves. To be specific, an external applied electric field of 5 V/nm can effectively weaken exchange interaction by 80% and is sufficient to induce nearly twofold change of the wavenumber. This discovery may open a door to energy-efficient local manipulation of the spin wave propagation utilizing electric fields, which is crucial for both fundamental research and spin wave based logic applications.Combined  Magnonics is deemed as the most promising candidate for information transmission and processing12357101112Benefited from the inspiration of CMOS technology where information is manipulated by electric fields with minimum power consumption, we present an energy-efficient local manipulation method for spin wave propagation utilizing electric-field-modulated exchange interaction in this article. The feasibility of electrical modulation on spin wave propagation makes CMOS-process-friendlier, lower-power-consumption and more compact magnonic devices possible in practical application.ab initio and micromagnetic simulations15ab initio calculations are carried out to investigate the direct influence of an external electric field on exchange constant based on density function theory. It turns out that the external electric field would induce a notable change in the exchange stiffness of ultrathin Fe film. Then micromagnetic modelings of spin waves propagating in thin film under influence of electric fields are implemented with parameters obtained by ab initio calculations. Our results show that the external electric field can effectively control the propagation of spin waves.Herein we carry out a multiscale and empirical parameter free calculation which combines −7 eV were employedFirst principles calculations based on density functional theory171820Our general supercell under investigation involves a stack structure of 3 bcc Fe(001)  monolaJ was extracted from the total energy difference between antiferromagnetic (AFM) and ferromagnetic (FM) configurations.The configuration under study was fully relaxed along the c-axis without electric field until the residual force worked out to be less than 1 meV/Å. The exchange integral F) and antiferromagnetic (EAF) energies:To facilitate description, we label the three Fe atoms as site 1, 2 and 3 from left to right. For simplicity, only the nearest neighbor exchange interaction is taken into consideration. Thus we have the following expressions for the total ferromagnetic  configuration in unit of ℏ.where Combining A can be related to the microscopic exchange integral in the Heisenberg model of ferromagnetism. For a body-centered cubic lattice, A is given byThe phenomenological parameter exchange constant a is the lattice constant and the exchange stiffness D is related to J bywhere A is calculated to be 1.85 × 10−11J/m, well in agreement with the counterpart (1.88 μerg/cm) estimated from the Curie temperature of bcc FeJ, exchange constant A and exchange stiffness D are listed in Results of our first principle calculations demonstrate that exchange constant do change under applied electric field, well in consistent with the experiment finding reported elsewhereIn the latter part of this article, we proposed a novel electrical modulation technique of spin wave length and phase velocity, using object-oriented micromagnetic frameworkMS the saturation magnetization, α is the Gilbert damping constant and γ is the Gilbert gyromagnetic ratio. where 3 in the numerical simulation. Parameters taken for simulation are: 1) saturation magnetization MS = 1.75 × 106 A/m; 2) Gilbert damping constant α = 0.03; 3) anisotropy constant K = 0 ; 4) Gilbert gyromagnetic ratio γ = 2.211 × 105m/(A·s); 5) DC bias field μ0H0 = 1 T. A stringent condition to follow in the creation of the meshes is that the maximum mesh element size should not exceed the exchange length of the material. The exchange length It is particularly pointed out that the exchange constant11 rad/s was applied to excite surface spin waveThe magnetization state for the configuration was first fully relaxed under the bias magnetic field along y direction. Then an excitation source varying as cosine function in time at a frequency of 4 × 10The micromagnetic simulation results in The electrical modulation of the wavelength can be interpreted by the spin wave dispersion relation. A useful analytic approximation for spin wave dispersion was derived by Kalinikos36ωex = ω0 + λexωMk2, ω0 = γμ0H0, ωM = γμ0Ms, d is the thickness of the thin film. ω-k view. For a given frequency of the spin wave this results in an upshift of the wavenumber k. It’s obvious that simulation result in Utilizing the dependence of the carrier wave number of the spin-wave for a given carrier frequency on the external electric field applied perpendicular to the ferromagnetic film, it is easy to implement electric-field-controlled phase shifter. Our findings will throw some light on the technical feasibility of microscopic spin-wave logic devices based on the electric-field-controllable manipulation of the spin wave phase7In conclusion, we have discovered the electric field control of exchange interaction in ultrathin magnetic film from first principle calculation and investigated its tuning effect on spin wave propagation by micromagnetic simulation. To be specific, an external applied electric field of 5 V/nm can effectively weaken exchange interaction by 80% and is sufficient to induce nearly twofold changes of the wavenumber. Therefore, this significant electric-field-modulation effect on the exchange constant is likely to pave another feasible route for the implementation of nanoscale spin-wave logic devices.How to cite this article: Wang, S. et al. Spin-wave propagation steered by electric field modulated exchange interaction. Sci. Rep.6, 31783; doi: 10.1038/srep31783 (2016).


Background. World health organization estimates that 25 million LBW babies are born annually worldwide and 95% occur in developing countries. Objective. To assess the prevalence and associated factors of low birth weight among term neonates delivered in Adwa Hospital, Northern Ethiopia. Methods. A cross-sectional study was conducted among neonates delivered in Adwa Hospital. All live births delivered from July 1, 2014, to June 30, 2016, were included in this study. The study participants were selected through systematic sampling technique and the data was collected using a structured questionnaire. Data was entered to Epi Data version 3.1 and analyzed using SPSS version 20 software. To identify independent predictors, bivariate and multivariable binary logistic regressions were employed. Adjusted odds ratio and 95% confidence interval were used to determine the strength of association. Results. The prevalence of term low birth weight was 10%. The risk factors were mothers aged less than 20 years, mothers whose pregnancy was desired, mothers with a history of abortion, and mothers with normal hemoglobin, iron with folic acid, and HIV status. Conclusion. The burden of LBW obtained in this study was in the same range as in some other countries.  Birth weight is the first weight of the foetus or newborn obtained after birth. Low birth weight refers to birth weights below 2500 grams . It is oMore than 20 million infants worldwide representing 15.5% of all births are born with low birth weight, 95.6% of them in developing countries . The preAs far as our knowledge, studies done on LBW and its risk factors are limited in Ethiopia. Therefore, the aim of this study is to assess prevalence of TLBW and its risk factors in Tigray region . This st2 and total population of 66826. It is 1004 km far from the capital city of Ethiopia, Addis Ababa, and 223 km from Mekelle, which is a capital city of Tigray. The study was conducted between July 1, 2014, and June 30, 2016. Cross-sectional study design was conducted at Adwa Hospital to collect the data.Adwa city is located in central zone of Tigray with an area of 23.0316 kmThe study population was sampled term newborn babies delivered at Adwa General Hospital for the period July 1, 2014, to June 30, 2016. Individual term newborn babies were reviewed from mother's chart or cards.Inclusion Criteria. Incorporated babies of all pregnant mothers delivered between 37 up to 42 week gestational age and singleton births were included. Exclusion Criteria. Babies born to mothers with medically known diabetes, babies born to mothers with hypertension, babies born to mothers with sickle cell hemoglobinopathy, records with incomplete data, postterm, preterm, and stillbirth were excluded. Sample size was calculated by using single population proportion formula (n = (za/2)2p(1 − p)/d2) based on the following assumptions: 95% confidence interval, 5% margin of error and taking 50% magnitude to get maximum sample size on term LBW, and 10% for incompleteness of secondary data (records). Hence, the sample size was 424 live births.20.5(1 − 0.5)/(0.05)2 = 385 adding 10% for nonresponse rate = 424 (1.96)Kth = 3000/424 = 7. We select randomly from the first seven listed participants and then we follow the same interval to allocate our sample size. The data extraction form is comprised of variables related to the maternal, neonatal, health related, and sociodemographic characteristics. The form was developed from World Health Organization, Ethiopian Demographic Health Survey, and other literatures. The data was extracted from Adwa General Hospital from two-year list of total term live deliveries obtained from delivery registration. The data was collected using data extraction form. This data collection procedure was carried out by five diploma midwives, and it was supervised by two BSc midwives. The data collector was oriented and trained for one day on how to extract the data, which records need to exclude, and about other data collection process. Brief introduction before and during data collection process was given.Two years mothers' cards were reviewed (37–42 wk of GA). The total mothers who gave live birth neonates were 3000 mothers within the two-year duration (2014–2016). The calculated sample size was 424. The method of sampling we used was systematic random sampling; after we make the complete list of the participants we calculate the interval Data quality was managed by recruiting BSc and diploma holders for each data collection and supervising working in the study site  based on experience of data collection and training was given for one day on how to collect and supervise the data using the prepared questionnaire. Pretest was done in 42 cards in Ayder Hospital before two weeks of actual data collection which was not included in the study and appropriate modification was made to the questionnaire after analyzing the pretest result like time modification, skipping pattern arrangement, wording, and phrases. Daily supervision, spot checking, and reviewing completed questionnaire was conducted to follow for completeness and consistency of the data (questionnaire) by the supervisors and over all data collection process was controlled by principal investigator.Data Management and Analysis. Data was coded and entered to computer using Epi Data software version 3.1 for its customizing skip benefit and finally data was tranfered to SPSS version 20 software package for analysis purpose. Data exploration was undertaken to see if there are odd codes or items that are not logical and then subsequent editing and cleaning were done before analysis.p value ≤ 0.25 at binary logistic were further analyzed in the multivariable binary logistic regression. Statistical significance for the association was considered at p value < 0.05. Binary and multiple logistic regressions were done to identify factors associated with term low birth weight neonate. Crude odds ratios were estimated for all independent variables in the binary logistic regression. All independent variables with p value less than 0.25 at bivariate analysis were entered into multivariate logistic regression to control for all possible confounders.Data was analyzed using SPSS version 20.0 computer software package. Data cleaning and editing were carried out. Frequency with percentage, pie charts, and bar graphs were used to represent results of categorical variables while mean (SD) and median (IQR) were used to represent continuous variables. Binary logistic regression  was used to see the strength of association between dependent variable and each independent variable. Finally, multivariable logistic regression was used to see the predictors of the outcome variable. Variables with Ethical Clearance. Ethical clearance was obtained from the Institutional Ethical Review Board of Mekelle University, College of Health Sciences. Confidentiality was maintained through anonymity in the use of the questionnaire. Medical card room workers were informed regarding the aim of the study. In order to maintain confidentiality of any information provided by the study subjects ID cards were used. Only the data collector and supervisor were present during the record extraction.Operational DefinitionsTerm low birth weight: a baby born after gestational age of 37–42 weeks with birth weight of less than 2,500 gramsObstetric health condition: the health condition of the mother during pregnancy and deliveryThe study involved a total of 424 mothers which is 100% of mother charts reviewed. The median (±IQR) age of the participants was 28 (±13 years); 272 (64%) of them were in the range of 20–34 years. 384 (90.6%) of the mothers were married and 219 (51.7%) of mothers' weights were above the mean. Slightly above half 241 (56.8%) of the study participants were rural residents. The majority of the mothers came from the Tigray community forming 391 (92.2%) and 183 (43.2%) of mothers were Para 2 to 4 see .Out of 424 mothers cards enrolled in the study 318 (75%) had a history of abortion and 402 (94.9%) wanted the pregnancy. 279 (65.8%) of them had ≥3 live children. Majority of them 384 (90.6%), 385 (90.8%), 397 (93.6%) took iron folic acid, had negative HIV test, and took tetanus toxoid immunization, respectively. 372 (87.7%) of the mothers were nonreactive with VDRL and 218 (51.4%) of the mothers had normal hemoglobin. 289 (44.6%) had less than four ANC follow-ups, and 350 (82.6%) had less than 2-year birth interval. A large number, 352 (87.4%), had no medical illness see .220 (51.9%) of the neonates were female; 321 (75.7%) of them had poor APGAR SCORES; 377 (88.9%) had no complications after delivery, and 400 (94.3%) had no congenital malformation see .The multivariable analysis of mother's data is shown in Table 4Age, history of abortion, pregnancy type, hemoglobin status, intake of iron/folic acid, and HIV status were significantly associated with term low birth weight while residence, maternal weight, marital status, parity, tetanus toxoid immunization, family planning, VDRL status, bad obstetric history, sex of neonate, and congenital anomalies were not associated with term low birth weight see .Mothers who delivered at the hospital at the age of less than 20 years were 1.7 more likely to deliver LBW babies than mothers aged 20–34 years . Mothers who wanted the pregnancy were 97% less likely to have LBW babies when compared to those mothers who had unwanted pregnancy . Mothers who had history of abortion were 2.4 times more likely to have LBW babies than those with no history of abortion . Mothers who had normal hemoglobin status were 98% less likely to give birth to LBW babies than those who had abnormal hemoglobin status . Mothers who took iron with folic acid were 99% less likely to have LBW babies than those who did not take iron and folic acid . Mothers who were reactive for HIV were seven times more likely to have LBW babies than nonreactive mothers  . The preThe prevalence of term low birth weight in this study was 10% which is similar to the result in Pakistan, 10.6% , but difMultivariate analysis shows that maternal age, pregnancy type, history of abortion, hemoglobin status, iron folic acid intake, and HIV status were significantly associated with term low birth weight. Also mothers less than 20 years of age delivering at the hospital were most likely to have term low birth weight, which is in agreement with studies done in Nepal  and IndiThe study shows that the type of pregnancy was significantly associated with term low birth weight. Newborns from unwanted and unplanned pregnancies were more likely than those from wanted and planned pregnancies to have LBW. This is similar to study done in Axum . These mPrevalence of term low birth weight in this study was almost similar to the results of other studies. The factors associated with term low birth weight are pregnancy type, maternal age history of abortion, hemoglobin status, iron folic acid, and HIV status. The burden of LBW obtained in this study was in the same range as in some other countries.


Nasolabial cysts are uncommon nonodontogenic lesions that occur in the nasal alar region. These lesions usually present with asymptomatic swelling but can cause pain if infected. In this case report, we describe the inadequacy of conventional radiography in a nasolabial cyst case, as well as the magnetic resonance imaging (MRI) and ultrasonography (US) findings in a 54-year-old female patient. A nasolabial cyst is a benign, slow-growing, nonodontogenic, primarily unilateral, extraosseous soft tissue lesion located in the nasal alar region below the nasolabial fold. The pathogenesis of nasolabial cysts is uncertain; however, there are two main theories. Some authors suggest that these lesions originate from displaced epithelium of the nasolacrimal duct remnants, while others suggest that it is a developmental fissural cyst originating from epithelial remnants entrapped between the lateral nasal, globular, and maxillary processes . These cNasolabial cysts cannot be seen on conventional radiography if there are no associated bone changes. However, these cysts may be aspirated and injected with a contrast agent for better visibility on plain radiographs , 5, 6. IThe purpose of this study was to report the case of a nasolabial cyst and to describe its USG and MRI exam features.A 54-year-old woman presented to our department with a common toothache. She described a history of a neurological examination with a brain MRI, and the neurologist sent her to us due to the possibility of a dentally originating lesion. During the clinical examination, we noticed a palpable fluctuant swelling on the upper labial sulcus, beneath the right nasolabial fold. There were no bony changes in the orthopantomography and occlusal radiography examinations, and all of the associated teeth were shown to be vital using electrical pulp testing . US was Based on the MRI, US, plain radiographic, and clinical findings, a preliminary diagnosis of a nasolabial cyst was made. The lesion was removed surgically with an intraoral approach under local anesthesia, and the surgical specimen was sent for a histopathological examination. The lesion was diagnosed as a nasolabial cyst.Nasolabial cysts have been known by many names. First described by Zuckerkandl in 1882, subsequent names appeared, such as Klestadt's cyst, nasal alveolar cyst, nasal wing cyst, and mucoid cyst of the nose; finally, Rao used the term “nasolabial cyst” as a more correct definition, which has remained in use until today . It is iNasolabial cysts are rare extraosseous developmental lesions that occur beneath the upper lip and adjacent to the alveolar process, above the apices of the incisors . They arFor the diagnosis of these lesions, several imaging methods can be used. Although plain radiographs may not show any detectable changes if there is no bone erosion, after the aspiration of the cystic fluid, the cyst can be injected with contrast material for better visibility , 5, 6. IUS is a diagnostic method that is often used for the examination of soft tissue lesions, and it can be used successfully in the maxillofacial region. US has some advantages over MRI and CT, like its nonionized characteristics and the fact that it is inexpensive . NasolabThe treatment of choice remains simple: enucleation with an intraoral sublabial approach and transnasal marsupialization, which has been performed with successful results. After careful and complete surgical treatment, recurrence is exceedingly rare .Nasolabial cysts are soft tissue lesions; therefore, conventional radiographs are often inadequate. Additional imaging modalities and clinical examinations are needed to diagnose them correctly. US and MRI are successful diagnostic imaging methods for evaluating the location, determining the contents of the cyst, and diagnosing, alongside a clinical examination. It may be advisable to use a combination of MRI and US in the diagnosis of nonosseous soft tissue lesions, such as nasolabial cysts.


It has long been thought that biomedical doctoral students pursue careers primarily as tenure-track/tenured faculty at research institutions. Recent reports showed, however, that the majority of biomedical doctoral alumni engage in a variety of careers. Wayne State University (WSU) undertook a project to understand the career trajectories of its biomedical doctoral alumni to create programs to better prepare its students for careers in multiple pathways. Data were collected on career outcomes of WSU’s biomedical doctoral alumni who graduated in a 15-year period from 1999–2014. Careers were classified into three tiers by Employment Sector, Career Types and Job Functions and career paths were examined by alumni gender, race, U.S. citizenship status, and association with certain academic characteristics. Several statistically significant differences in career paths among all demographics were found. For example, women were more likely to be in teaching and providing healthcare, men in faculty and research; Black alumni pursued careers in Government at higher rates and Whites in For-Profit careers; Asians and non-U.S. citizens spent more time in training positions than others. There was no association of academic characteristics such as GRE, GPA, and Time-to-Degree completion with careers in the two largest sectors of Academia or For-profit. Since our trainees are engaged in this rich variety of careers essential to advancing biomedical science and research nationally, it is imperative for the graduate training community to embrace all careers as successful, and transform the model for biomedical doctoral training to foster student success across this broad career spectrum. Academic institutions have long held the belief that biomedical doctoral students pursue careers primarily as tenure-track/tenured faculty in research institutions, and therefore training programs focused almost exclusively on preparing students for academic careers. Until recently, study section reviewers of training grants from federal agencies such as the National Institutes of Health (NIH) recognized academic careers as the primary outcome of success, further reinforcing this belief. Recent reports regarding career outcomes of doctoral alumni, however, showed that, in reality, almost 75% of all biomedical doctoral students engage in a variety of careers beyond academia, including for-profit, government, and non-profit sectors , 2. It iWith increasing calls for transparency of reporting career outcomes –6, some Additionally, very few reports addressed demographics of students engaged in these various careers , 14, resWayne State University (WSU) is a comprehensive research institution with an enrollment that includes 1,500 doctoral students in 75 doctoral programs. Like other academic institutions, we also operated under the assumption that our doctoral students engage in careers almost exclusively in academia. Our training models were designed to prepare students solely for academic careers. To ensure that we were accurately understanding the short- and long- term career pathways of our Ph.D. recipients, we collected career outcomes data of doctoral alumni who graduated over a 15-year window from 1999 to 2014 [We used the career in which the alumnus was engaged at the point in time of the data collection and “binned” these data in aggregate for alumni based on number of years from graduation . These aggregated career outcomes data were then classified per the unified three-tier taxonomy , and useIn 2015 WSU’s Graduate School launched an Alumni Census Project in which the current employment information of 866/950 (91%) biomedical doctoral alumni who graduated from 1999–2014 from biomedically-related programs were collected, as previously described . These pThis project was conducted with approval from Wayne State University’s Institutional Review Board on the Use of Human Subjects, IRB#094013B3E.All data are reported in aggregate or with identifiable information removed. Data in which a group is below 4% are not reported in order to maintain confidentiality and anonymity of the individual(s).The 865 alumni include 459 women (53%), 406 men (47%); 464 White (54%), 334 Asian (39%), 48 Black (5%), 19 total Native American, Pacific Islander race (2%); 478 U.S. citizens/permanent residents (55%), and 387 non-U.S. citizens (45%), as shown in i.e., Window 1 (0–5 years), Window 2 (6–10 years), and Window 3 (11–15 years) immediately following graduation. Note that in this manuscript, the trajectory of each alumnus over a 15-year time period is not reported. Rather, the overall alumni aggregate employment data are shown in each time window from years following graduation within each of the three tiers at the specific time of the survey. So, for example, a student who graduated in 2006 would be represented in the 6–10 year block only. Career outcomes are reported in percentages; numbers from which these percentages are derived are presented in Tables A-X in 2) analyses with follow-up z tests employing a Bonferroni correction were used to test for significantly different proportions of alumni in different Employment Sectors, Career Types, and Job Functions over time. Since time windows contained different sets of participants, between-subjects analyses were conducted. Multinomial logistic regression analyses were then conducted to test for significant interactions between time windows and demographic characteristics. However, none of these interactions were statistically significant and in some cases , singularities in the Hessian matrix due to small sample sizes in some cells prevented valid multinomial analysis. In the absence of significant interaction effects and because of the Hessian matrix violations, Chi-Square (Χ2) analyses were used to test for significantly different proportions of demographic groups within each tier. As with time windows, post hoc z tests with Bonferroni corrections were used to test for significant effects if the omnibus Chi-square test was significant. Differences among comparison groups were considered to be statistically significant at p < .05. Statistical output tables are reported in Tables A-L in 2) analyses were also attempted to examine patterns of career outcomes in isolated subsets of alumni; however, small and n = 0 cell sizes for some categories resulted in uninterpretable results. Because the patterns of findings appear to be similar for these small groups, we decided to present analyses on each demographic variable (rather than combinations of variables) to yield robust results that could be used as a basis for future investigations. Thus, the analyses presented here focus on patterns of career outcomes within each demographic group for the entire 15-year window.Outcome analyses were performed using SPSS version 25 (IBM 2018). Chi-Square (ΧAcademic characteristics assessed were (a) GRE-Quantitative and GRE-Verbal scores; (b) cumulative GPA at doctoral graduation; and (c) time to doctoral degree completion. Average Time-to-Degree at WSU for biomedical doctoral students is 5.5 years. Logistic regression analyses were used to test whether academic characteristics were associated with Employment Sector outcomes. Significance was determined with a p value < .05.In our overall alumni outcomes, not all employment sectors, career types and job functions were equally represented  = 20.30, p = .002. Post hoc z tests were performed to determine when alumni representation increased or decreased in particular sectors. While the majority of alumni start out in Academia, more alumni are represented in sectors outside of Academia, especially when comparing Window 1 to Window 3, p < .05. One reason for this drop is that the Academic sector includes postdoctoral and other trainees (19%), and as expected, they move from traineeships to other sectors. The only other significant change is the increase over time in the number of alumni who were employed in the For-Profit sector (p < .05). Very small percentages of alumni pursue careers in Government and Non-profit sectors and no statistically significant changes in alumni representation occur for these sectors.Almost 90% of alumni are engaged in either Academia or For-profit sectors. The pattern of Employment Sector changed over time, ΧThere are no significant differences between men and women in any Employment Sectors, with the trends being the same as those in the total population.2  = 20.17, p = .003. A higher proportion of Asians entered Academia compared to Whites (p < .05) whereas a higher proportion of Whites compared to Asians chose careers in For-profit (p < .05). The only other significant difference is seen with Blacks in that they are employed in significantly higher percentages in the Government sector as compared with Asians or Whites (p < .05).There was a significant effect of Race on Sector ΧThere were no significant differences between U.S. citizen and non-U.S. citizen alumni showed similar patterns in Employment Sectors, with a small increase in the For-profit sector for non-U.S. citizens from Window 1 to Window 3.2  = 116.05, p = .0001. As expected and desired, there was a significant decline in the percentage of alumni in Further Training after Window 1 . There is a significant increase in the number of alumni in Science Related careers between Windows 1 and 2 (p < .05) and an increase in the number of alumni in careers Not Related to Science from Window 1 to 3 (p < .05).The majority of alumni are in Primarily Research and Science Related Career Types, with a smaller percentage in Primarily Teaching. A small percentage of alumni is in careers Not Related to Science. The distribution of Career Types changed over time, Χ2  = 22.72, p = .0001. Significantly more women were in Primarily Teaching careers (p < .05), and more men in Primarily Research careers (p < .05).Men and women differed in their Career Types, Χ2  = 47.27, p = .0001, were accounted for by Asians pursuing Further Training in higher proportions compared to White and Black alumni (p < .05), and Whites choosing Career Types that were Not Related to Science at a higher rate than Asians and Blacks (p < .05).Racial group distributions, Χ2  = 55.94, p = .0001. U.S. citizens were more likely to be in Primarily Teaching, Not Science-Related, or Science-related careers . Significantly more non-U.S. citizens were employed in Further Training and Primarily Research careers (p < .05 for both).Career Type was associated with citizenship status, Χ2  = 120.72, p = .0001. As expected, the proportion of alumni who start in Postdoctoral or related traineeships immediately following graduation is followed by declines over the time windows (p < .05). More alumni are represented in Faculty and Technical Director or Group Leader positions after Window 1 (p < .05).There were significant changes in the proportions of alumni in Job Functions over the time windows, Χ2  = 16.36, p = .003 was accounted for by the fact that women choose careers as Healthcare Providers at a significantly higher rate than men (p < .05).The significant Chi-square for gender, Χ2  = 39.26, p = .0001, with differences between Asians and Whites seeming to account for this overall effect. Asians were more likely to be Postdoctoral (Scientific Research) scholars compared to Whites (p < .05) whereas White alumni were significantly more likely to be Faculty and Healthcare providers (p < .05 for both comparisons).Significant racial group differences in Job Functions also emerged, Χ2  = 51.40, p = .0001, were due to a significantly higher percentage of non-U.S. citizens entering Postdoctoral (Scientific Training) (p < .05), as well as higher percentages of U.S. citizens pursuing careers as Health Care Providers and Faculty (p < .05 for both comparisons).Significant citizenship effects, ΧWe examined the association of academic characteristics with the two largest Employment Sectors of Academia and For-profit, since participation in other sectors is too small to allow for meaningful comparisons. Please note that not all students take the GRE since it is not a requirement for admission to all biomedical programs at WSU and therefore the numbers do not total 865. Data are expressed as % alumni in each score/year range.There are no statistically significant differences in any of the academic characteristics examined between alumni in Academia and For-profit sectors. In response to national calls for transparency –3, 15, WAbout 92% of our biomedical doctoral alumni are employed in careers related to research and science. This finding counters current literature that argues that the PhD pipeline is producing too many graduates for not only academia, but also for industry and government . While wOverall we find that our alumni do not necessarily remain in a single Employment Sector. We do not know if this change of employment areas is purposeful or is influenced by other decisions such as financial, family, and need to remain in certain geographical locations, among others. Regardless, it is clear that for whatever reason, alumni distribution in careers changes over time, emphasizing the mobility  and the Gender does not correlate with differences in Employment Sectors, but men and women show interesting differences in Career Types and Job Functions. More women engage in Primarily Teaching careers and as Healthcare Providers, while more men engage in Primarily Research careers. While we do not know the reasons for these differences, recent discussion in the literature implies that women prefer jobs providing more flexibility such as teaching and health care over jobs in academia because of family obligations and work-life balance –24.Race is associated significantly with many outcomes among our alumni in all three tiers. For the Employment Sector, the largest differences observed are that Asian alumni enter Academia at higher rates compared with Whites and Blacks; Black alumni are more likely to be in the Government sector compared with Asians and Whites; and White alumni are more likely than Asians to engage in For-profit sectors. While no race-based differences are found among Primarily Research, Primarily Teaching or Science Related Career Types, more Asians tend to remain in Further Training compared with Blacks and Whites, and White alumni choose Not Related to Science careers at a higher rate than Asians or Blacks. For Job Functions, most significant differences are seen among Asian and White alumni. Our White alumni pursue careers as faculty and Healthcare Providers at higher rates than Asians, and Asians remain for longer durations in Postdoctoral Training compared with Whites. Race and ethnicity patterns among biomedical doctoral alumni have been described in several studies –29, and There is a dearth of literature on the career outcomes of U.S. citizens and non-U.S. citizens. A recent study shows that significantly more U.S. postdoctoral alumni pursue careers in for-profit ventures and obtained more tenure/tenure-track positions than international postdocs . At WSU,Regardless of gender, race, and U.S. citizenship status, our data show that alumni engage in a variety of careers. Academic institutions, faculty, and funding agencies should respect these multiple career pathways because scientists and researchers in these areas contribute to and enhance scientific discovery in new and meaningful ways. Our alumni shape public policy in government jobs, lead research and innovation efforts in industry, influence public perception of science through communications, and teach the next generation of science researchers and practitioners. We must embrace these careers as successful outcomes of doctoral training and make it acceptable for students to explore and identify their career interests in programs developed by the graduate training community, such as the NIH-funded Broadening Experiences in Scientific Training (BEST) programs  includinOne of the most important findings from our study is that alumni in the two largest employment sectors of Academia and For-profit show very similar academic characteristics. There is no statistically significant difference between the two groups in either GRE scores, cumulative GPA, or Time-to-Degree completion. This debunks the myth, for our institution at least, that students who have “better” academic characteristics engage in faculty careers while “other” students settle for careers in other sectors. These data should empower students entering doctoral programs to have honest discussions about their career aspirations with their research advisors and to explore careers outside the professoriate, if they so choose.In summary, WSU’s 15-year doctoral biomedical alumni are engaged in a variety of careers. Differences are seen among careers of alumni by gender, race, and U.S. citizenship status. There is no correlation of academic characteristics examined of alumni with careers in Academia and For-profit sectors. In addition to disciplinary research training, it is imperative for the graduate training community to embrace all science and research-related careers as successful, and transform training models such that biomedical doctoral students are provided with the appropriate professional development skills to succeed in a broad spectrum of careers. Modernizing curricula and providing opportunities for experiential learning and engaging the private sector is also required to better align education with career paths and empower students to make informed career decisions , 19, 32.S1 Table(DOCX)Click here for additional data file.S2 TableTables A-H in S2 Table for Employment Sector. Table A: Total alumni data for Figs (DOCX)Click here for additional data file.S3 TableAnalyses are shown in Tables A-D for Employment Sector ; Tables E-H for Career Type ; and Tables I-L for Job Function . Neither GRE-Q nor GRE-V significantly predicted the likelihood of entering the Academic or For-Profit Employment Sector .(DOCX)Click here for additional data file.S1 File(PDF)Click here for additional data file.


Women who were exposed to tobacco smoke every day delivered neonates with significantly lower mean birth weight (mean ± SE: 2703 ± 135 g) than women who were only exposed once a week (mean ± SE: 3125 ± 147 g) (p < 0.05). A one-minute increase in cooking time in a kitchen without a chimney increased women’s expired air carbon monoxide concentration by 0.038 ppm (p = 0.006). Long-term exposure to wood-fuel smoke in a kitchen without a chimney can increase the risk of inhaling high concentrations of carbon monoxide.Low birth weight is a key public health problem in many developing countries, including Sri Lanka. Indoor air pollution from tobacco smoke and kitchen-fuel smoke are among the major contributors to low birth weight, factors of which there are little awareness of in Sri Lanka. We evaluated the effect of passive smoking and kitchen-fuel smoke exposure on birth weight. Seventy-six pregnant women were included in the study. Data were collected by questionnaire, and exposure assessment was conducted using a breath carbon monoxide monitor. Women exposed to second-hand tobacco smoke daily had a significantly lower mean gestational age at delivery (mean ± standard error [SE]: 38.0 ± 0.5 weeks) than women who were exposed to second-hand tobacco smoke only once a week (mean ± SE: 39.3 ± 0.3 weeks) ( Low birth weight (LBW) is defined as body weight less than 2500 g at birth. The World Health Organization estimates that more than 20 million LBW infants are born each year, and that LBW affects approximately 16% of all newborns in developing countries. The prevalence of LBW in Sri Lanka has been 16–17% for many years ,4,5,6. AMany low- to middle-income countries still use biofuels  as the main source of energy for cooking and heating. A number of studies have demonstrated a clear association between solid-fuel use and lower respiratory tract infection and chronic obstructive lung diseases ,12,13. HSri Lanka is a developing country which has recently been termed a lower- to middle-income country. Wood is the main source of cooking fuel in Sri Lanka, and is often burned over an open fire in an inefficient cooking stove. In addition, liquid petroleum gas and kerosene are used in some households. To our knowledge, this is the first study in Sri Lanka to evaluate the effects of second-hand tobacco smoke and wood-fuel smoke exposure on neonatal birth weight in relation to measured levels of expired CO.This was part of a larger study carried out to identify maternal factors associated with neonatal birth weight. A prospective study was carried out in a tertiary care hospital in Sri Lanka between October 2015 and June 2016. A more detailed description of the subject recruitment procedure has been published separately . In brie+ Smokerlyzer breath CO monitor . The breath CO monitor is accurate within ±2 ppm and it was calibrated at least every six months in line with the manufacturer’s recommendations. CO and COHb were each measured once only during the clinic visit. Women were asked to hold their breath for 15 s and then blow in to the instrument. Breath CO was measured in parts per million (ppm), and blood COHb was measured as percentage of oxygen replaced. CO levels were defined as equivalent to non-smoker (0–4 ppm), danger zone (5–6 ppm), smoker (7–10 ppm), and frequent smoker (11–16 ppm). The corresponding COHb levels were defined as 0–1.27, 1.43–1.59, 1.75–2.23, and 2.23–3.19%, respectively [Eighty-seven pregnant women who visited the hospital antenatal clinic at 30 weeks of gestation were included in this part of the study. Data were collected using a pre-tested questionnaire that included questions on socioeconomic and demographic factors, exposure to tobacco smoke during the pregnancy, the frequency of exposure, relationship to the smoker, use of wood fuel for cooking, time spent cooking, and the characteristics of the kitchen. During the same clinic visit, maternal expired air CO concentrations and percentage of COHb were measured using a piCOMaternal blood hemoglobin concentrations at the booking visit (weeks 6–8 of gestation) and weeks 28–30 of gestation were obtained from individual pregnancy cards, whereas neonatal birth weight and gestational age at birth were obtained directly from the hospital records after deliveries. Trained research assistants administered the questionnaire and took measurements.The study was approved by the ethical review committees of the Graduate School of Health Sciences at Niigata University, Japan (No. 125), the Faculty of Allied Health Sciences, University of Peradeniya, and the teaching hospital at Kurunegala, Sri Lanka (ERC/2015/06). The study was conducted in compliance with the principles of the declaration of Helsinki. Written informed consent was obtained from the participants. t-test were used to compare groups exposed and not exposed to tobacco smoke and wood-fuel smoke. Correlations between continuous variables  and neonatal birth weight were evaluated using Pearson’s correlation coefficient. To test the effects of categorical variables  on birth weight, the study used one way ANOVA followed by multiple regression analysis. The variables with p < 0.2 in the ANOVA procedure were subsequently included in the multiple regression analysis as dummy variables in addition to the continuous variables mentioned above. Simple regression analysis was used to test the effects of wood-fuel exposure time on mean CO levels. R2 (adjusted for the number of predictors in the model) was used to show how much variance is explained by the models. Statistical analysis was carried out using Minitab version 17. Chi-squared test and two-sample At the end of the data collection period, neonatal data were missing for 14 subjects because of ambiguity in hospital delivery registry records, as multiple similar maternal names made it difficult to accurately locate the study participants. The final analysis therefore consisted of maternal-neonatal data for 76 mother–child pairs.n = 26), 69.2% had been exposed inside the house (closed space) and the others outside. Daily tobacco smoke exposure was reported by 61.5% and 38.5% reported that they were exposed approximately once a week. Of the women exposed to tobacco smoke, 69.2% had been exposed because their husbands smoked and 26.9% because of the father or father-in-law.p = 0.012). Pre-pregnancy BMI and maternal age had no significant correlation with birth weight (p > 0.05).Estimates of Person’s correlation coefficient showed that there was a moderate positive correlation between birth weight and gestational age at delivery . Women in the highest income category  delivered babies with significantly higher mean birth weight than women in the lowest monthly household income category (p = 0.013). Mean birth weight of the babies of primiparous mothers was 258 g below those of multiparous mothers (p = 0.045)  .p > 0.05). Univariate analysis showed that women exposed to tobacco smoke on a daily basis had a significantly lower gestational age at delivery than women who were exposed once a week . They also delivered babies with significantly lower mean birth weight . When the analysis was restricted to women with a kitchen chimney, the group exposed tobacco-smoke on daily basis still had significantly lower mean neonatal birth weight than the group exposed only once a week  and COHb  among women exposed to wood-fuel smoke. The correlation was strong when the analysis was restricted to women who were exposed in a kitchen without a chimney . Regression analysis found that a 1-min increase in cooking time in a kitchen without a chimney increased expired CO by 0.038 ppm  and COHb by 0.006%   .We found that women who were exposed to daily tobacco smoke and women exposed to wood-fuel smoke in a kitchen without a chimney had higher levels of breath CO and COHb. The lack of significance might be because the biomarkers were measured several hours after the women left home. CO has a shorter half-life, so a gap of a few hours from exposure to measurement would reduce the level considerably. Exposure to motor vehicle exhaust fumes on the way to the clinic might also interfere with the estimate of the CO readings. A study in Peru found higher levels of CO in the air in the kitchens of bio-fuel users . HoweverOur findings are based on a small cohort, but this is one of the few prospective studies of which we are aware that assesses personal CO levels during pregnancy against exposure to second-hand tobacco and biomass fuel smoke. None of the women enrolled in our study were active smokers, as hardly any Sri Lankan women smoke. To date, no Sri Lankan studies have been published on second-hand smoke exposure during pregnancy. We found a significantly lower mean birth weight and lower gestational age in babies whose mothers were exposed to tobacco smoke during pregnancy on a daily basis compared with once a week. We also found a moderate positive correlation between gestational age and birth weight. The higher birth weight is probably a result of longer gestation, as the fetus is able to gain additional weight as a result. All the women who were exposed to second-hand tobacco smoke were also exposed to wood-fuel smoke. Second hand tobacco smoke could therefore have a confounding effect on the relationship between wood fuel smoke exposure and birth weight, and wood fuel smoke exposure could act as a confounding variable in the relationship between second hand tobacco smoke exposure and birth weight. It was difficult to test these confounding effects due to difficulty in carrying out a meaningful multiple regression analysis with this smaller sample size. Hackshaw  reportedp > 0.05). Longer cooking times may account for the higher expired CO concentrations. In addition, the time since last exposure and the point of data collection is different for each woman may be one of the reasons behind unexpected CO values. A number of studies ,18,19 hap > 0.05). The lowest birth weight was reported in babies whose mothers cooked in a kitchen without a chimney and were exposed to second-hand tobacco smoke (mean ± SE: 2588 ± 353 g), but the comparison was not statistically significant  on the way to the clinic, which could affect the accuracy of the CO readings. We did not assess the time gap between the most recent exposure and the point of data collection, and this may cause biased results because of the short half-life of CO. In addition, we were limited to the primary fuel type, although some women may use firewood in combination with liquid petroleum gas and/or kerosene.Despite the limitations, this is the first study of which we are aware in which the effects of second-hand tobacco smoke and wood-fuel use on neonatal birth weight have been assessed among Sri Lankan women. It is also the first prospective study to our knowledge that provides quantitative exposure response data on neonatal birth weight in relation to measured levels of CO in a Sri Lankan context. Data from this small study can be used to design further large scale studies in Sri Lankan context to assess the effects of smoke exposure on human health. We recommend future studies to assess exposure to smoke among pregnant women and its effects on neonatal birth weight, addressing some of the limitations of this study.Second-hand tobacco smoke exposure in Sri Lankan pregnant women is followed by low mean birth weights of their babies. Long-term exposure to wood-fuel smoke in a kitchen without a chimney can increase the risk of inhaling high concentrations of CO.


Background: Patterns of adverse birth outcomes vary spatially and there is evidence that this may relate to features of the physical environment such as air pollution. However, other social characteristics of the environment such as levels of crime are relatively understudied. This study examines the association between crime rates and birth weight and prematurity. Methods: Maternity inpatient data recorded at birth, including residential postcode, was linked to a representative 5% sample of Scottish Census data and small area crime rates from Scottish Police forces. Coefficients associated with crime were reported from crude and confounder adjusted models predicting low birth weight (< 2500 g), mean birthweight, small for gestational age and prematurity for all singleton live births. Results: Total crime rates were associated with strong and significant reductions in mean birth weight and increases in the risks of both a small for gestational age baby and premature birth. These effects, with the exception of prematurity, were robust to adjustment for individual characteristics including smoking, ethnicity and other socio-economic variables as well as area based confounders including air pollution. Mean birth weight was robust to additional adjustment for neighbourhood income deprivation. Conclusion: The level of crime in a mother’s area of residence, which may be a proxy for the degree of threat felt and therefore stress experienced, appears to be an important determinant of the risk of adverse birth outcomes. There are plausible hormonal pathways linking the two including levels of certain hormones such as placental corticotropin-releasing hormone (CRH) and adreno-corticotropin hormone (ACTH). Both are important mediators in the pathway towards elevated risks of experiencing both premature birth and small for gestational age (SGA) babyA fairly large number of studies investigating stress and adverse birth outcomes have focused on exposure to acute stressors finding associations with events such as earthquakes,Other studies focused on chronic contextual stressors such as incidents, rates and perceptions of crime in local neighbourhoods. The possibility of crime rates explaining spatial patterns in general health outcomes has been exploredHowever, there remains a lack of consistent international evidence linking local environment sources of chronic or ambient stressors such as crime rates to adverse birth outcomes that is adequately adjusted for potentially confounding effects such as air pollution for example. In this study we test for an association between potentially stressful neighbourhood environments in Scotland, measured through small area based total crime rates in mother’s place of residence, and various adverse outcomes including prematurity and foetal development. We hypothesise that the levels of recorded crime in the mother’s immediate local area of residence will be associated with outcomes at birth.The Scottish Longitudinal Study (SLS), a 5% sample of the Scottish population, linking census records and a number of other administrative data sets,The sample was restricted to singleton births (born between 1994 and 2008 inclusive), those that contained complete, non-missing information and those with a valid postcode for the mother’s usual place of residence when the birth was registered. We derived the following outcomes; risk of moderately (32–36 weeks) and very (< 32 weeks) preterm birth, risk of small for gestational age , risk of low birthweight (LBW < 2500 g) and mean birth-weight. For the models predicting prematurity and SGA all singleton live births were included but for the other outcomes premature births (less than 37 weeks) were excluded.,The crime domain of the Scottish Index of Multiple Deprivation (SIMD) was used to estimate mothers’ ‘exposure’ and was linked to the main sample through residential postcode. The SIMD crime domain records incidents of crime from police forces across Scotland. This measure expresses the total number of crimes of violence; domestic house breaking; drugs offences; minor assault; and vandalisms as a rate per 10 000 population at a datazone level (6505 in Scotland with each zone designed to contain between 500 and 1000 households). Given the relatively higher frequency and spatial correlation of vandalism, drug offences and minor assaults, this measure will tend to be high in areas where these events are high. The measure is therefore capturing areas that are visibly confrontational and potentially violent. Previous studies have demonstrated that these types of areas in particular have a direct and independent influence on individuals fear of crime.We used the 2006 and 2009 releases of the SIMD which contain crime statistics from 2004 and 2007/2008, respectively. Exposure to levels of crime was determined by taking the centroid of the maternal residential postcode recorded in the maternity hospital record and linking this to the corresponding datazone. Datazones in city centres were excluded as they contained disproportionately higher rates of crime associated with night-time type urban activity.2), nitrogen dioxide (NO2) and particulate matter smaller than 10 microns (PM10)].Confounding is a problem when analysing associations between a spatial exposure such as area crime rates and birth outcomes because the latter is strongly associated with other social and residential characteristics of the mother that may also be risk factors. The relative importance of individual and area factors and their association with health has been a matter of debate for many decades, both generallyFinally, adverse birth outcomes such as infant mortality are associated with area deprivationMultilevel logistic, linear and multinomial regression models were used to explore the relationship between different birth outcomes and neighbourhood crime rates. The crime variable was log-transformed because the various crime measure distributions were highly positively skewed. We tested a number of different functional forms of varying degrees, including non-integer polynomial functionsWhen testing the sensitivity of these results to adjustment for area income deprivation , many ofFrom a nationally representative sample of pregnant mothers in Scotland, this study has found that area rates of crime in the maternal area of residence are associated with large and significant reductions in mean birth weight and increases in the risks of both a small for gestational age baby and premature birth. These associations, with the exception of prematurity, were robust to adjustment for a range of individual characteristics including smoking and ethnicity as well as area based estimates of pollution concentrations. However, with adjustment for area income deprivation only mean birth weight was significantly associated with area crime rates.,,et al.Our findings appear to broadly corroborate results from the relatively few previous studies in this area.Only one other study examined different crime types and was restricted to an examination of the risk of preterm birth stratified by race.There is evidence to suggest that chronic rather than acute experiences are the more epidemiologically important forms of stress which is perhaps due to the accumulation of physiologically damaging effects over longer periods of time. This process adds to an individual’s ‘allostatic load,’,In the main models in this study, we did not, unlike other studies,There are a number of possible explanations or interpretations for this effect. It could be that there is an upper threshold limit for the effect of external environmental exposures beyond which additional restriction to a baby’s attainable weight is impossible. This interpretation is premised on the notion that baby weights in the lower ranges of the birth weight distribution are likely to be to mothers who been exposed to multiple individual and environmental risk factors which act in combination to greatly restrict the baby’s attainable weight. Amongst these babies, it is possible that exposure to additional environmental exposures such as neighbourhood crime rates, has a reduced effect because their attainable weight is sufficiently low that greater restriction is impossible. Thus, in the sensitivity models, the crime variable competes with the area income deprivation variable for this limited weight restricting environmental effect. Conversely, babies whose attainable weight is closer to the overall mean of the birth weight distribution may be more susceptible to additional environmental insults, particularly the chronic forms of stress that may be associated with living in environments characterised by high rates of crime. The higher attainable weights in this area of the distribution allows for greater scope for additional exposures to affect birth weight with this greater variation allowing crime to retain predictive power when including the collinear predictor of area income deprivation. This might explain why the effect of exposure to areas with high crime rates is greater for mean birth weight than for LBW and SGA; because of a greater biological susceptibility among babies born to mothers with lower exposures to other risk factors.an independent area effect, including it would constitute over-adjustment.Another interpretation could be that additional adjustment for area income deprivation constitutes incorrect adjustment for a variable which is not a true confounder of the relationship of interest; in other words that it constitutes over-adjustment.This study has limitations. Firstly, the available crime information was restricted to two years which were towards the end of births in the cohort and we therefore assume that crime levels remained relatively stable over time. However, if they do not, the resultant measurement error is most likely to be at random and therefore bias the effect towards the null. Secondly, uncertainty in the length of gestation which in our data, as in many others, was derived from the mother’s estimated date of last menstrual period may have introduced error in the models which will reduce the precision of resulting estimates.In conclusion, this study has found that adverse birth outcomes, particularly those related to foetal development, are associated with the rate of recorded crime in the mother’s area of residence at birth. This crime measure is likely to particularly reflect high levels of vandalism, minor assault and drug offences and therefore captures environments visibly laden with threat that elicit heightened fear of crime among residents.


Perinatal mortality according to birth weight has an inverse J-pattern. Our aim was to estimate the influence of familial factors on this pattern, applying a cohort sibling design. We focused on excess mortality among macrosomic infants (>2 SD above the mean) and hypothesized that the birth weight-mortality association could be explained by confounding shared family factors. We also estimated how the participant’s deviation from mean sibling birth weight influenced the association.We included 1 925 929 singletons, born term or post-term to mothers with more than one delivery 1967–2011 registered in the Medical Birth Registry of Norway. We examined z-score birth weight and perinatal mortality in random-effects and sibling fixed-effects logistic regression models including measured confounders  as well as unmeasured shared family confounders (through fixed effects models). Birth weight-specific mortality showed an inverse J-pattern, being lowest (2.0 per 1000) at reference weight (z-score +1 to +2) and increasing for higher weights. Mortality in the highest weight category was 15-fold higher than reference. This pattern changed little in multivariable models. Deviance from mean sibling birth weight modified the mortality pattern across the birth weight spectrum: small and medium-sized infants had increased mortality when being smaller than their siblings, and large-sized infants had an increased risk when outweighing their siblings. Maternal diabetes and birth weight acted in a synergistic fashion with mortality among macrosomic infants in diabetic pregnancies in excess of what would be expected for additive effects.The inverse J-pattern between birth weight and mortality is not explained by measured confounders or unmeasured shared family factors. Infants are at particularly high mortality risk when their birth weight deviates substantially from their siblings. Sensitivity analysis suggests that characteristics related to maternal diabetes could be important in explaining the increased mortality among macrosomic infants. Birth weight is a strong predictor of stillbirth and neonatal mortality. The relation between birth weight and mortality is an inverted J-pattern with a steep mortality slope among those small at birth (the long arm), and a shorter slope among high birth weight infants (the hook) ,2. SinceEpidemiological –8 and biUsing family data to separate individual and family effects of birth weight on perinatal outcomes is a way to potentially illuminate some of these explanations ,38. ThisSince 1967, the MBRN has recorded data on all births in the country . The natDetails of the establishment of the study population are provided in The Regional Committee for Medical Research Ethics approved the study.The study outcome was perinatal death, defined as stillbirth (death before or during delivery), or early neonatal death (death within one week after delivery) . AdditioThe main exposure variable was birth weight. In order to compare all participants and siblings, a z-score birth weight was standardized for sex, parity, gestational week, year of birth, and plurality, into a zBW variable with mean = 0 and standard deviation (SD) = 1. We categorized zBW into 10 ordered levels, as well as a broader categorization of infants into microsomic (zBW <–2), normosomic (zBW –2 to +2), and macrosomic (zBW >+2).We constructed four variables based on the mothers’ births: “mean sibship zBW”, “deviance from mean sibling zBW”, “paternity”, and “sibship size”. We defined mean sibship zBW as the mean zBW of all the mother’s infants, categorized into four levels. Deviation from mean sibling zBW was defined as the difference between a participant’s zBW and mean zBW of all other infants to the mother. We ordered this variable into five categories: ≥1 SD lower than sibling mean zBW; 0.5 to 1 SD lower; <0.5 SD difference; 0.5 to 1 SD higher; and ≥1 SD higher. The middle three categories were collapsed into a “<1 SD difference” category in some analyses. The paternity variable was based upon the father’s identity for infants in the maternal family. The main purpose was to assess if the paternal contribution to birth weight in the sibship came from one or more fathers. We applied three categories: “same father” if all infants with zBW data in the maternal families had the same father, “not same father” if some infants with zBW data had different fathers, and “uncertain” in the 1500 (0.2%) maternal families with a combination of the same father and unidentified fathers. Sibship size was a count of all infants to the mother.We considered a number of other covariates based on their potential to influence the zBW-PNM association –14, suchFactors considered to have a potential of mediating a causal zBW effect –14,23 wei are mediating factors by which birth weight could cause death during delivery. This mediating pathway could only apply to stillbirths during delivery and early neonatal deaths. For stillbirths before delivery, the same factors could be common effects (colliders) of birth weight and death [In order to explain our analytic choices, we made a causal diagram where the roles of the covariates with respect to the relation between zBW and PNM were depicted . Here, tnd death . As an end death  rather tThe main analytical aim was to estimate individual-level associations between perinatal death and zBW, with emphasis on macrosomia, taking sibling zBW into account.rowsort command. Descriptive characteristics of zBW, the covariates, and their relations to birth weight and perinatal death were calculated in ordinary tabular analysis.We used Stata/SE 13.1 software . The zBW variable was created by z-transforming and standardizing birth weight for sex, parity (5 categories), gestational week, year of birth, and plurality (2 categories) using Stata’s In the analyses of individual-level associations between death and zBW, we applied two analytical procedures. Both included mother’s identity as the grouping variable. First, we applied random intercept panel-data logistic regression in the entire study population or in selected population subgroups. Second, we implemented a conditional (fixed-effects) logistic regression in the 7686 families (23 546 infants) with discordant perinatal survival experiences. The fixed-effects approach is essentially a conditional analysis estimating differences within sibships, and, contrary to the random-effects models that only control for factors included in the model, all shared (family) factors are invariant and controlled for by design. The advantage of fixed-effects is balanced by lower statistical power. Both procedures yielded risk coefficients (ln OR) and 95% confidence intervals (CI). Because we had emphasis on macrosomia, we ran analyses both with 10-level zBW and zBW where the three macrosomic zBW levels >+2 were collapsed. The zBW category (+1 to +2) with the lowest crude PNM served as reference.We fit three random-effects models and one fixed-effects model. Details of the four models are outlined in In this way, we were able to control for measured individual confounders and for measured and unmeasured shared confounders. This left the role of unmeasured individual confounders unexplained. The role of mediators was assessed for early neonatal death only.We conducted analyses both in the total population and in subgroups. The most important was a comparison of early (1967–1998) and more recent (1999–2011) births, because aforementioned potential confounders were only measured in the more recent years.The initial analyses revealed that associations between perinatal death and macrosomic categories were considerably stronger in pregnancies recognized as diabetic compared to non-diabetic pregnancies. We investigated this in a random-effects analysis with an interaction term of diabetes and zBW categories.Differences in associations in mutually exclusive subgroups were tested using the approach of Altman and Bland . We consTo obtain estimates of the modifying strength of sibling zBW on the zBW-PNM association, we computed a 30-level interaction variable combining the 10-level zBW variable and the three-level deviance from mean sibling zBW variable. We substituted this interaction variable for the zBW and the mean sibship zBW variable in models where the potential confounders in Model 2 were included. Our choice of reference was zBW +1 to +2 and <1 SD different from sibling mean zBW. This analysis was mainly done to examine if deviance from mean sibling zBW influenced mortality risk among macrosomic infants, but also to confirm studies that have shown that small and normal-sized infants have an added risk when they are of smaller size than their siblings –43.punaf procedure, we were able to compare the observed population PNMO with the hypothetical PNMH that would have been experienced if all had had unchanged zBW category but the same PNM risk as that of groups with similar (<0.5 SD different) zBW as their siblings. The punaf procedure yielded PAF estimates = 1 –PNMH/PNMO with 95% CI [The impact of sibling zBW on the zBW-PNM relation in the total population and selected subgroups was assessed by estimating PAFs. This was done in unconditional logistic regression models that included the zBW categories, the 5-level deviance from mean sibling zBW variable, and the potential confounders from Model 2. Using Stata’s h 95% CI .Three issues were considered in sensitivity analyses.First, we assumed that the deviance from mean sibling zBW variable was a proxy for the individual zBW difference from an unknown programmed optimum birth weight. Accordingly, we did not differentiate between elder and younger siblings . HoweverSecond, we assumed that the difference from mean sibship zBW variable was an indicator of deviance of own birth weight from an unknown programmed optimum birth weight, but only an imperfect one. We expected that this indicator would be better for the 726 584 families with the same mother and father than for the whole population, because of the father’s genetic contribution to offspring birth weight. We also assumed that deviance from mean sibling zBW would be a better indicator for large (>2 siblings) than for small (1–2 siblings) families. These indicator quality analyses were performed by comparing PAFs in the subsets, assuming larger impact for full sibling families and large families.Third, we assumed that conditions related to maternal diabetes and metabolic disturbances were underreported in the MBRN, furthermore, that a similar interaction could be expected for such unrecognized conditions as for the observed diabetes variable. We investigated the impact of such underreporting by creating a hypothetical diabetes variable with higher prevalence than the observed one. This was achieved by changing status from non-diabetic to hypothetical diabetes for chosen fractions of the population. This procedure was done randomly within each specific zBW-PNM stratum, securing that zBW distribution and zBW-specific PNM were kept similar to the observed diabetes variable. The random-effects associations were then compared for observed diabetes and the more prevalent hypothetical diabetes variables.We recorded 8115 perinatal deaths (PNM 4.2 per 1000); 5775 (70%) being stillbirths and 2440 early neonatal deaths. Only 751 of stillbirths occurred during delivery, 3731 before delivery, and the remaining 1193 had unspecified timing. The relation between zBW category and PNM in The synergistic pattern of maternal diabetes and zBW on PNM is shown in the left-hand columns of vs 0.563) was significant  and the ratio of the two ORs (2.9/1.8) was 1.7 (95% CI 1.1 to 2.4). The explanation was a different secular trend in PNM across zBW categories: PNM for microsomic infants and normosomic infants was strongly reduced between the two periods, while PNM among macrosomic infants remained the same (4.5 in 1967–1998 and 4.4 in 1999–2011). This also meant that the PNM fraction related to macrosomia rose from 0.023 in 1967–1998 (160/6815) to 0.055 in 1999–2011 (71/1300).Associations for macrosomia were estimated in the confounder-adjusted Model 2 in which zBW categories >+2 were collapsed . MacrosoThe crude Model 1 coefficient for macrosomia was 0.786 . The attenuation from Model 1 and the confounder-adjusted Model 2  was 12%.We also examined the macrosomia–PNM association in the fixed-effects analysis among the 23 546 infants in the 7686 discordant families .smaller than their siblings, and for macrosomic infants larger than their siblings. The same analysis restricted to sibships with the same father . PNM was 2.5 times higher for those with zBW >1 SD higher than the elder sibling (PNM 4.0) than for those with similar (<1 SD different) zBW (PNM 1.6).O, and on the hypothetical PNMH under the assumption that all infants had the same mortality risk as those with small (<0.5 SD) difference from sibling mean zBW. PNMO and PNMH were 4.2 and 3.0, respectively  had as result that the majority of macrosomic deaths changed from observed non-diabetic to hypothetic diabetic category  confounders, and mediators. Maternal diabetes and macrosomia had an interaction (synergistic) effect on PNM. All covariates had only moderate impact on the inverted J shape and the PNM association with macrosomia. Contrary to the study hypothesis, shared confounders, as evaluated in fixed-effects analysis, did not explain the hook of the inverted J. By contrast, deviance from mean sibling zBW modified the zBW–PNM relation: the relation between macrosomia and mortality was stronger if the infant was larger than their siblings compared with the magnitude of the relation if the infant was a similar size as siblings. Similarly, the relation between microsomia and mortality was stronger if the infant was smaller than their siblings compared with the magnitude of the relation if the infant was a similar size as the siblings. This is the first study to demonstrate that the modification of PNM risk by difference in size between siblings is a general pattern that covers the entire birth weight range. The study is also the first ever to provide a quantitative estimate of the influence of deviance between own and sibling birth weight on PNM, indicating that family (sibling birth weight) is associated with a substantial portion of the variance of the birth weight–mortality relation.In this large population-based study, we linked siblings in consecutive birth records in a procedure independent of maternal recall. Birth records included data on a number of relevant covariates. MBRN registration includes all infants born in Norway .The participants constituted 92% of singleton term or post-term births to mothers with more than one birth in the study period. Exclusions were mainly due to missing gestational age data.Model misspecification could be a problem in this observational study. We governed the analyses according to the causal diagram . PreviouInformation bias could pose a problem. We consider data on zBW and perinatal death to be reliable, rendering the crude zBW–PNM association reasonably valid. Associations with macrosomia could be underestimated because this is a heterogeneous entity with a complex etiology . QualityWe were able to account for measured confounders and unmeasured shared confounders, but not for confounding by unmeasured individual factors. This could pose an important threat to valid inferences. We lacked data on potentially important pregnancy-specific factors. Examples are placenta function and metabolic characteristics that could have profound effects on birth weight and survival, and thereby have major influence on the results. We consider incomplete data on BMI and particularly maternal diabetes to be a limitation. The prevalence of diabetes in the MBRN barely reached 2% in the most recent years. This is in contrast to a many-fold higher occurrence of reported gestational diabetes in more recent studies . The preSeveral covariates in this register-based study were crude approximations , and we would expect that adjustments in the multivariable models for this reason would be incomplete. Incomplete recording of some covariates could add to this problem. Except for maternal marital status and sibship size, we had no indicators of parental socioeconomic position. This probably poses a minor problem since socioeconomic status seems to have little influence on the relation between birth weight and mortality in countries such as Norway . We wereLack of power constituted a problem in small zBW strata, particularly for the macrosomic categories where PNM risk was lower than in the microsomic area. The same applies to some of the subset analyses and to the fixed-effects analysis because <1% of families were discordant with respect to perinatal deaths.When assessing the generalizability of the results, one should keep in mind that the study was restricted to singleton and term or post-term births, and that it related to a relatively homogenous and low-risk population in a developed country. The study covered an extended period and results could therefore be outdated. The results for births between 1999 and 2011 suggest however that the findings are still relevant.th-95th percentile but a strong association for >95th percentile weights ([The inverse J-pattern of the mortality curve is soundly documented –8. The hweights (, Table 3Evolutionary biologists interpret the inverted J-pattern on the population level a result of natural  or stabiThe results suggest that factors causing deviance from optimum birth weight in the individual birth are pregnancy-specific and not shared by siblings. Numerous studies show that maternal pregestational and gestational diabetes mellitus have impact on macrosomia and perinatal mortality –18. ThisOur results underline the importance of obstetricians taking size at birth in the family into account when assessing mortality risk. Apart from obstetrical trauma, perinatal death among macrosomic infants could be heterogeneous and more of a problem in pregnancies complicated by poor glycemic control. If so, management in pregnancy according to recent guidelines  could prMacrosomia affects only a small proportion of all births, and only 3% of all perinatal deaths affected this weight category in the present study. Its public health impact is however increasing because macrosomia worldwide has become more common . FurtherThis sibling-designed study shows that family factors have large impact on the relation between birth weight and survival. Deviation of birth weight from sibling birth weight in either direction is associated with excess PNM. In our study, the excess among macrosomic infants was not confounded by shared family factors as hypothesized, but was evident among those who outweighed their siblings. The main explanation of the hook of the inverted J is probably found among unknown pregnancy-specific factors. Future studies should identify and include more complete and refined data on such factors, metabolic and others. Causal paths including macrosomia and perinatal mortality are intricate, and models that are more refined can probably be achieved by applying causal analytical methods .