Several studies have reported on associations of size at birth and early growth with general and central obesity; however, few have examined the potential effects of birth weight and postnatal growth on separate abdominal fat compartments. We investigated the effects of size at birth, linear growth and relative weight gain from birth to adulthood on visceral (VFT) and subcutaneous abdominal (SAFT) fat thicknesses at age 30 years.Z-score changes, conditional relative weight gain and conditional height at several ages.A total of 2663 participants from the 1982 Pelotas (Brazil) birth cohort study had complete information on ultrasound measures of abdominal fat at age 30 years, and anthropometric measurements for at least five visits (0/2/4/23/30 years). We estimated weight and height P⩽0.01). Women born with intrauterine growth restriction (IUGR) had greater VFT than other women , and they showed a stronger positive influence of infant weight gain 0–2 years on VFT . Stunting at 2 years was associated with lower SAFT but not VFT, and it modified the influence of weight gain 2–4 years on SAFT in both sexes (both Pinteraction<0.05).In both men and women, VFT and SAFT showed positive associations with conditional relative weight gain during all age periods beyond 2 years and birth, respectively (all Our findings reinforce the advantages of being born with an appropriate birth weight, and the hazards of rapid postnatal gains in weight relative to linear growth, particularly after the critical window of the first 1000 days. An unfavourable intrauterine environment can prevent the foetus from attaining its growth potential and result in adaptive problems. This process – so-called ‘intrauterine growth restriction' (IUGR)In the present study, we aimed to: (i) test the hypotheses that higher weights and faster growth in childhood would increase abdominal fat in adults from a middle-income setting and (ii) investigate whether the above trajectory is modified by IUGR or nutritional status at age 2 years.20 Briefly, it recruited 5914 live births in 1982 to mothers living in the urban area of Pelotas, a city in Southern Brazil. The first follow-up in 1983 targeted only children born between January and April 1982; 1457 children were examined (~80% of the target sample). In 1984 and 1986, all urban households were visited, resulting in follow-up rates of 87% (n=4934) and 84% (n=4742), respectively. In 1997, 70 census tracts  were randomly selected and visited to locate 1076 cohort members (72% of the target). In 2000, 2250 males (78.9% of the target) were interviewed and had physical examinations at compulsory Army recruitment. In 2001, all households in the 70 census tracts selected in 1997 were revisited, locating 1031 women (69% of target). In 2004–2005, another household census was performed and 4297 cohort members (77% of target) were interviewed and examined. Most recently, in 2012–2013, 3711 cohort members attended a research clinic follow-up (68.2% of target). In the early phases, verbal informed consent was obtained from the mothers, whereas in recent phases, written consent was obtained. Study protocols were approved by the Ethical Committee of the Federal University of Pelotas.19The 1982 Pelotas Birth Cohort Study has been previously described.Trained anthropometrists collected the measurements following standard protocols. Infants were weighed to the nearest 10 g using calibrated pediatric scales . Children were weighed to the nearest 0.1 kg using portable mechanical scales . Adolescents and adults were weighed to the nearest 0.1 kg using calibrated electronic scales .Birth length was not recorded. Supine length and height was measured to the nearest 0.1 cm in 1984 and 1986, respectively, using boards manufactured locally according to international specifications . From 1997 onwards, standing height was measured to the nearest 0.1cm using a full-length wall-mounted stadiometer .Z-scores by comparison with the 2006/2007 World Health Organization (WHO) growth standards.21 Age and sex-specific internal Z-scores were calculated for weight and height at ages 18/19, 23 and 30 years, excluding pregnant or post-partum women. We combined data from 2000  and 2001  as follow-up at age 18/19 years. IUGR was defined as BW for gestational age and sex below the 10th centile using the reference developed by Williams et al.22 Stunting (height Z-score below –2) and overweight (BMI Z-score above +2) at age 2 years were defined by 2006 WHO criteria.21BW and childhood measurements were converted to age and sex-specific 24 Briefly, visceral fat thickness (VFT) was estimated by the distance between the peritoneum and the lumbar spine at the intersection between the xyphoid line and the waist circumference. Subcutaneous abdominal fat thickness (SAFT) was estimated at the same probe site by the distance between the posterior line of dermis and the outer bowel wall. Both distances were measured from static images taken at the end of a quiet expiration when applying minimal pressure. Women with known or probable pregnancy, or up to 3 months post-partum, were excluded from this assessment. The relative intra-observer technical error of measurement for the visceral thickness was 4.1% and 3.4% for subcutaneous fat thickness, whereas the relative inter-observer technical error of measurement was 3.1% for both measurements.Abdominal ultrasound was performed using a Toshiba Xario  ultrasound machine with a 3.5-MHz convex probe following validated protocols.a priori potential confounders were collected.20 Family income in 1982 was calculated as the sum of the monthly incomes of all working persons living in the household, expressed in multiples of the minimum wage (⩽1/1.1–3/>3). Gestational age was calculated by mother's recalled last menstrual period. For 522 study members, gestational age was missing or invalid; for 507 of these, multiple imputed values (five imputations) were calculated using a regression model containing maternal age, skin colour, height, smoking during pregnancy, offspring sex and family income at birth.25 Maternal education was assessed as completed years of formal education, and was categorised into three groups (0–4/5–8/9+ years); maternal self-reported skin colour was categorised into two groups (white/non-white); maternal smoking during pregnancy was categorised into five groups (non-smokers/1–14 cigarettes during part of pregnancy/1–14 cigarettes during whole of pregnancy/15+ cigarettes during part of pregnancy/15+ cigarettes during whole of pregnancy). Maternal height was measured and pre-pregnancy weight was obtained from the antenatal care register or reported by the mother and used to calculate pre-pregnancy BMI. Missing BMI values were imputed using the same method as described for gestational age.25Data on Statistical analyses were performed using Stata version 13  and all analyses were stratified by sex. VFT and SAFT were transformed to normal distributions using logarithmic and square root functions, respectively. Both variables were posteriorly standardised to allow direct comparisons of the regression coefficients for these outcomes.Growth was analysed separately by sex, in subjects with complete anthropometry and ultrasound data, excluding women who were pregnant or post-partum in 2000, 2004–2005 and 2012–2013. In all 2 men and 22 women had undergone abdominal plastic surgery, but when these individuals were excluded from the analyses the results remained unchanged (data not shown). Therefore, we present the results on the whole available sample.28 Conditional relative weight gain was estimated by regressing the weight at each age on all previous weight and height measures, and also concurrent height. Conditional height was estimated by regressing the height at each age on all previous weight and height measures, but not concurrent weight.26Conditional growth models were performed by generating conditional weight or height measures that are uncorrelated with all previous measurements and therefore indicate the influence of recent growth rate (since the preceding time point) on the ultrasound outcome measure.Z-scores for BW and attained weight and height at ages 1, 2, 4, 15, 18–19, 23 or 30 years; conditional relative weight gain at ages 2, 4, 23 or 30 years; and conditional relative height gain at ages 4 or 23 years. Additional analyses were carried out for shorter age ranges based on the subsamples measured at ages 1, 15 and 18/19 years. Models were adjusted for family income at birth, maternal education, maternal skin colour, maternal height, maternal BMI before the pregnancy, smoking in pregnancy and gestational age.Multiple linear regression models tested the associations with VFT or SAFT and β) coefficients indicate the standardised difference in VFT or SAFT per +1 s.d. of the exposure. To aid interpretation, some coefficients were back transformed using the inverse function and expressed in centimetres (cm). Baseline models for changes in weight or length/height Z-score were adjusted for weight or length/height Z-score at the beginning of the period.Beta  with regard to sex, maternal skin colour and education, and prevalence of IUGR and overweight at 2 years; however, the selected sample had slightly higher family income at birth (P=0.004) and lower prevalence of low BW . This sample was similar to the original cohort , but lower SAFT ; both differences persisted after adjustment for BMI .Anthropometric measurements are summarised in P<0.001). Maternal height showed a strong positive association with SAFT, but only among males; in addition, we found weak but significant positive associations between maternal BMI before pregnancy and both fat compartments in both sexes. Smoking during pregnancy was inversely related to both VFT and SAFT, but only in women , but no difference in SAFT (P=0.33). By contrast, men born with IUGR had lower adult SAFT than other men , but no difference in VFT (P=0.35). Regarding nutritional status at age 2 years, men and women who were stunted had lower adult SAFT than other individuals , whereas those who were overweight at age 2 years had higher SAFT  than in non-IUGR girls . Furthermore, conditional weight gain (between 2 and 4 years) showed stronger positive associations with SAFT in non-stunted men and women than in stunted individuals .We conducted interaction analyses in order to test if IUGR, stunting and overweight at 2 years modified the associations between early conditional weight gain and adult VFT or SAFT. In fully adjusted models, 3 out of 16 interaction tests were significant at the 5% level . Early cThe findings from this large, long-running, middle-income country birth cohort study suggest differences in the relative contributions of childhood weight gain and linear growth to the two abdominal fat compartments, VFT and SAFT, in adults. In both men and women, higher relative conditional weight gain during most ages from infancy to adulthood was associated with greater VAT and SAFT. Moreover, the associations of weight gain early in life with abdominal fat appeared to be modified by IUGR (girls only) and stunting at age 2 years. In women, shorter attained height from infancy to adulthood was associated with higher VFT, whereas height gains were positively associated with SAFT.26 Nevertheless, few studies have investigated such growth patterns in relation to the abdominal fat mass.13 Several statistical approaches have been described to identify growth trajectories or critical age periods related to later disease risk.29 In our study, the conditional growth measures represent the acceleration or deceleration in growth from an individual's previous trajectory.Growth patterns throughout life have been associated with obesity risk in adulthood; the most consistent associations are with rapid growth during infancy and early childhood and with early BMI rebound.β coefficients for these associations strengthened substantially with age, which is consistent with the progressive increase in adiposity that begins at the BMI rebound period (at 3–7 years). A one s.d. increase in conditional relative weight gain from 2 to 4 years predicted a small increment (0.2 cm) in VFT in adulthood in both sexes, whereas the same increase in weight gain from 4 to 23 years predicted a more substantial increment (0.9 cm on average). We should point out that a one s.d. increase in conditional relative weight gain represents a large change in weight trajectory, and the estimated coefficients on adult VFT are small given the wide range in this measurement (2–16 cm).Our results suggest that rapid weight gain (independent of changes in height) promotes the accumulation of both VAT and SAFT, in both sexes; no critical windows were obvious –associations were apparent with weight gain at all ages from 2 years onwards for VAT, and at all ages from birth onwards for SAFT. However, the 30 However, when we considered BW as a categorical variable, IUGR in women was associated with higher adult VFT, and a positive association between infant weight gain (0 –2 years) and VFT was seen in IUGR but not in non-IUGR women. Nevertheless, we found no evidence of interaction between IUGR and later rapid weight gain (after 2 years of age) on VFT or SAFT. These findings may seem contrary to previous studies that suggested a protective role of early rapid weight gain in the first 2 years of life  against overall adiposity, in comparison with adverse consequences of later rapid weight gain (at 2–4 years of age).31 However, our current findings with these specific abdominal fat measures, in particular VFT, may have stronger relevance to metabolic disease risk.8 Furthermore, the accumulation of visceral and subcutaneous adipose tissues may involve different physiological mechanisms,33 and in light of the substantial recent increase in the prevalence of obesity in our cohort (from 8.5% in 2004–2005 to 23.6% in 2012–2013), it may prove to be more beneficial to gain subcutaneous abdominal fat instead of visceral fat.We found no association between BW and adult abdominal fat mass, a finding that is consistent with the few studies, all set in high-income countries, that used direct imaging methods.34 Our results suggest that stunting at 2 years may limit the accumulation of SAFT but not VFT in both sexes; an ongoing preference to accumulate VFT rather than SAFT might be deleterious to metabolic health. In support of this idea, we observed that women with shorter height had more VFT, but unfortunately our lack of data on linear measurements at birth meant we were unable to distinguish between antenatal and early postnatal growth retardation.Previous studies have reported that linear growth retardation in early childhood is associated with short stature and less lean mass in adulthood; however, such studies have not specifically measured abdominal fat.19 The lower prevalence of low BW in subjects measured at 30 years of age is largely due to the fact that low BW infants in the original cohort were more likely to die and, therefore, less likely to be available for the follow-up at 30 years of age. In addition, we cannot rule out the possibility that selective survival of a subgroup of low BW newborns  may partly explain the association between IUGR and VFT in women. Second, many participants had missing values for gestational age and maternal pre-gestational weight and, therefore, we used a multiple imputation approach. Third, our main results are restricted to subjects with complete information on growth measurements at birth and at least five follow-up examinations. Further measurements at ages 1, 15 and 18/19 years were available in only small subsamples, and therefore we had limited ability to determine associations with specific age periods of growth, such as the BMI rebound and weight gain during adolescence. Finally, we relied on ultrasound measurements of VAT and SAFT as proxies for these abdominal fat masses; however, previous studies assessing the validity of the same protocol reported strong correlations with magnetic resonance imaging estimates of abdominal fat in a variety of settings and populations.36 A strict quality control process was carried out and we could identify consistent sex differences in the distribution of these abdominal fat compartments compared with previous reports using other imaging methods.39We acknowledge the other limitations of our study. First, cohort members who attended the latest assessment tended to be wealthier and less likely to have low BW; these differences can be partly attributed to survival bias, as deaths in early life were more common among the poor and low BW children.40We therefore consider that our study represents an advance on previous knowledge regarding the early determinants of abdominal obesity in our cohort, by assessing separately the visceral and subcutaneous abdominal fat compartments. Moreover, we applied complementary methods to analyse growth trajectories associated with these outcomes, including relative growth and weight gain, which allows assessment of age-specific conditional measures independent of all previous weight and height measures, takes into account the strong correlation between repeated measurements over time and avoids the ‘reversal paradox'.In conclusion, rapid weight gain in different age periods, but particularly after the age of 2 years, was associated with greater adult abdominal fat. In addition to rapid weight gain, markers of growth restraint, IUGR and short stature, appeared to confer greater susceptibility to accumulating visceral fat in women and influences abdominal subcutaneous fat in both sexes. Previous analyses from this cohort have been shown to be consistent with those in other birth cohorts set in lower- and middle-income countries. Thus, we consider that our findings point to metabolically adverse combinations of early growth restraint and rapid weight gain in populations in these settings. In terms of policy implications, our findings reinforce the advantages of being born with an appropriate BW, and the hazards of rapid postnatal gains in weight relative to linear growth, particularly after the critical window of the first 1000 days.


Subjects born preterm have an increased risk for age-associated diseases, such as cardiovascular disease in later life, but the underlying causes are largely unknown. Shorter leukocyte telomere length (LTL), a marker of biological age, is associated with increased risk of cardiovascular disease.To compare LTL between subjects born preterm and at term and to assess if LTL is associated with other putative cardiovascular risk factors at young adult age.We measured mean LTL in 470 young adults. LTL was measured using a quantitative PCR assay and expressed as T/S ratio. We analyzed the influence of gestational age on LTL and compared LTL between subjects born preterm (n = 186) and at term (n = 284). Additionally, we analyzed the correlation between LTL and potential risk factors of cardiovascular disease.r = 0.11, p = 0.02). Subjects born preterm had shorter LTL (mean (SD) T/S ratio = 3.12 (0.44)) than subjects born at term (mean (SD) T/S ratio = 3.25 (0.46)), p = 0.003). The difference remained significant after adjustment for gender and size at birth (p = 0.001). There was no association of LTL with any one of the putative risk factors analyzed.Gestational age was positively associated with LTL (Young adults born preterm have shorter LTL than young adults born at term. Although we found no correlation between LTL and risk for CVD at this young adult age, this biological ageing indicator may contribute to CVD and other adult onset diseases at a later age in those born preterm. Nowadays, 5–13% of all newborns in developed countries are born preterm  : AlreadyTelomeres are noncoding repetitive DNA sequences at the end of each chromosome. Their primary function is to maintain genomic stability ,12. Due Studies reporting on the relation between telomere length and preterm birth are scarce and contradictive. A small study n = 26) measured LTL in umbilical cords of neonates and found no difference between LTL in preterm and full-term controls . Henckel6 measureWe hypothesized that accelerated biological ageing may at least partly explain the increased risk of CVD in subjects born preterm and used LTL as an ageing biomarker to test this hypothesis. We therefore investigated the correlation between gestational age and LTL and analyzed differences in LTL between subjects born preterm and at term. Additionally, we assessed if, at the age of 21 years, putative risk factors for later CVD  correlate with shorter LTL.The study population consisted of 470 healthy individuals, aged 18–24 years ,24. SubjThe Medical Ethics Committee of Erasmus Medical Centre approved the study. Written informed consent was obtained from all participants.Participants were invited to visit Erasmus University Medical Centre. Prior to the visit, participants fasted for at least 12 hours and abstained from smoking and alcohol for at least 16 hours.Birth data regarding gestational age and birth size were obtained from hospital records, primary health care records and general practitioner records. Information regarding socioeconomic status (SES) and smoking of the participants was obtained using questionnaires. Education level of the participants was used as socioeconomic indicator to determine SES . Height After centrifugation, all blood samples were kept frozen until assayed (-80°C). For measurement of hsCRP, an important predictor of future atherosclerotic events, an in-house-high-sensitivity ELISA with polyclonal rat CRP antibodies for catching and tagging  was used. Total cholesterol level was measured using the CHOD-PAP and the GPO-PAP reagent kit . High-density lipoprotein (HDL) cholesterol level was measured using a homogeneous enzymatic colorimetric assay (Roche Diagnostics). Low-density lipoprotein (LDL) cholesterol was calculated using the Friedewald formula: LDL cholesterol level in mmol/L = total cholesterol–HDL cholesterol level– 0.45 x level of triglycerides.Genomic DNA was isolated from peripheral leukocytes using standard procedures. All LTL measurements were made in the same laboratory at the University of Leicester, without knowledge of birth status. Mean LTL was measured by the quantitative PCR-based technique as previously described ,29. Telohttp://www.growthanalyser.org).Standard deviation (SD)-scores for birth length and birth weight were calculated in order to correct for gestational age and gender . SD-scorBaseline characteristics of normally distributed data are presented as mean (SD) and of non-normally distributed data as median (interquartile range (IQR)). ANOVA (continuous data) and Chi square tests  were used to determine differences between participants born either preterm or at term. The association between gestational age and LTL was determined using multiple linear regression analysis. After assessing the linear correlation (Model A), adjustments were made for age, gender, birth length SDS, birth weight SDS and adult height SDS (Model B). The interaction term birth length SDS * adult height SDS was added to the analysis because the study group had been selected on birth length and adult height, in order to ensure that the effect of these variables was modeled correctly. Additionally, we adjusted for smoking and SES (Model C). Difference in LTL between subjects born preterm and at term was analyzed with an independent samples t-test. In an ANCOVA model, we additionally corrected for gender, birth length SDS and birth weight SDS. The association between LTL and risk factors of CVD was assessed with linear regression, with LTL as dependent variable and the different risk factors as independent variables. Finally, we assessed if there were differences in these risk factors between subjects in the bottom quartile for LTL and subjects in the upper quartile for LTL, using independent samples t-tests.Results were considered statistically significant if the p-value was <0.05. Statistical package SPSS version 21.0  was used for all analyses.Characteristics of the study population are shown in 2 = 0.01, p = 0.02). Adding age, gender, height SDS, birth weight SDS, birth length and the interaction term birth length SDS * adult height SDS to the model , improved the model and the correlation between gestational age and LTL remained significant (p = 0.03). Also, a positive association was found between female gender and LTL (p = 0.002). Gestational age and gender both remained significant after adding smoking and socioeconomic status  to the analysis .We evaluated the relative contribution of several variables at birth and at young adult age to LTL in a multiple regression analysis . There wWe assessed the relation between LTL and several putative CVD risk factors, including insulin sensitivity (Si), at 21 years of age . There wIn this study, we analyzed the association between gestational age and LTL in 470 young adults and compared LTL between subjects born preterm and at term. We found that gestational age is positively associated with LTL and that subjects born preterm have shorter LTL than subjects born at term. This difference remained significant after correction for birth length and birth weight, indicating an independent effect of gestational age on LTL, not confounded by birth size. We also found that females had longer telomeres than males, which is in concordance with earlier studies . At thisper se is not a risk factor for shortening of telomeres. In contrast to our study, telomere length was measured in saliva in that study, which questions the comparability of their results to our findings. Kajantie et al. [Thus far, there is limited literature on the association between preterm birth and LTL and the results are contradictive. Friedrich et al.  measurede et al.  describePrevious studies have highlighted oxidative stress as an important determinant of LTL ,33 and sTo provide a more meaningful context in terms of kilobases for the observed difference in T/S ratio between those born preterm and those born at term, we used data from a previous study from the same laboratory where a comparison had been made between LTL measured by PCR and in kilobases by Southern blotting ,41. On tPreviously, we have shown that several putative risk factors for CVD are already increased at a young age in subjects born preterm ,4. We thIn conclusion, our data show that gestational age is positively correlated with LTL and that young adults born preterm have shorter LTL than young adults born at term. This could reflect pre- and postnatal oxidative stress and in turn could partly explain the association between preterm birth and later life risk of CVD. Since the prevalence of preterm birth and survival is rapidly increasing, our results are of clinical relevance for a large and increasing number of subjects worldwide.S1 TableValues are given as mean (sd). DBP = diastolic blood pressure; HDLc = high-density lipoprotein cholesterol; hsCRP = high sensitivity C-reactive protein; LDLc = low-density lipoprotein cholesterol; SBP = systolic blood pressure; Si = insulin sensitivity; TC = total cholesterol; Tg = triglycerides.(PDF)Click here for additional data file.


To study the characteristics of orbital cyst associated with microphthalmos in a group of Chinese patients, and to analyze the relationship between orbital cyst and orbital volume.Cross-sectional comparative study.120 patients who were diagnosed as unilateral clinical blind microphthalmos, in which 20 patients had orbital cyst in the affected eye.Participants had computed tomography (CT) scan. CT images were analyzed with a computer-aided software.Volume and position of orbital cyst, microphthalmic to contralateral ratio (MCR) of orbital volume, height and depth and orbital rim displacement.38.1% of the cysts located anterior to or at the equator of the globe, 75% of which located infratemporally and all of which were outside the muscle cone. Most (84.6%) of the posterior cysts were inside the muscle cone. The anterior cysts were larger than the posterior cysts (p = 0.005). MCR of orbital volume (p<0.001), height (p = 0.004) and width (p = 0.043) were significantly higher in patients with orbital cyst than controls. For patients with orbital cyst, larger cyst-plus-globe volume of the affected eye was associated with higher MCR of orbital volume . Patients with large cyst-plus-globe volume had higher MCR of orbital volume (p = 0.002), height (p = 0.014), width (p = 0.005) and depth (p = 0.002) and less displacement in inferior (p = 0.004) orbital rim, compared with patients with small cyst-plus-globe volume, but the differences between patients with small cyst-plus-globe volume and patients without cyst were not significant.Microphthalmic eyes with large cyst-plus-globe volume showed better similarity with the contralateral eyes, comparing with microphthalmic eyes without orbital cyst or with small cyst-plus-globe volume. It suggested that the existence of orbital cysts  in microphthalmic eyes might play a positive role in the orbital development. Microphthalmos with orbital cysts is a rare, severe malformation resulted from defective closure of the embryonic fissure and over-growth of the inner layer of the optic cup. There have been many case reports about this congenital anomaly since its first report in 1858. McLean eOne of the most important esthetic problems of microphthalmic patients is the hypoplasia of the affected orbit and hemifacial deformity. In our clinical practice, we found that some of the microphthalmic patients with orbital cysts showed less retardation of development in the affected orbit, compared with patients without cyst. It occurred to us that the orbital cysts, which serve as volume replacement and simulate the orbital implants to some extent, might also play a positive role in orbital expansion. In our previous study, computerized tomography (CT) and iPlan Cranial software were used to measure orbital parameters and allowed us to evaluate the orbital development in Chinese children with congenital microphthalmos. With CT This clinical comparative study included 120 patients who were diagnosed as unilateral clinical blind microphthalmos in Beijing Tongren Hospital between July 2009 and January 2015 in a consecutive manner, in which 20 patients were found to have orbital cyst in the affected eye. All the patients had finished CT scanning. The study was approved by the Ethics Committee of Beijing Tongren Hospital and adhered to the provisions of the Declaration of Helsinki for research involving human subjects. Written informed consent was obtained from all adult participants. For the minors/children, written informed consent was obtained from their parents.Clinical blind microphthalmos was defined as 1) markedly decreased ocular size to the contralateral eye / markedly decreased axial length to normal axis adjusted for age ; 2) lack of normal globe structure on imaging examinations ; 3) no light perception. Exclusion criteria were secondary microphthalmos  and diseases that need to be differentiated from microphthalmos with orbital cyst .Examinations included visual acuity, general slit-lamp examinations, CT scan, and a detailed questionaire. Each of the newly diagnosed microphthalmic patients was consulted by a pediatrician to find any systemic abnormalities. Data were collected before any treatment.Each patient was scanned with a Brilliance 64-channel multidetector CT scanner  with the orbitomeatal line as baseline. Scanning parameters for adults were 120kV, 250mA, 16×0.625mm detector collimation, with a pitch of 0.563. Images were reconstructed on a 512×512 pixel matrix at a thickness of 0.67 mm. To minimize the radiation exposure in children, we adhered to the ALARA (as low as reasonably achievable) principle by adjusThe bony orbit was outlined in each slice of axial CT scans. The anterior border of the orbit was defined as the line drawn between the inner aspects of the orbital rim on each side. At the level of medial canthi, the line was drawn between the lateral rim and the anterior lacrimal crest. The posterior border was a line that enclosed the foramens and fissures of the orbit. 3D volume rendering was implemented and orbital volume was calculated .The globe was outlined in each slice of axial CT scans and the global volume was calculated by the iPlan Cranial software The cyst was outlined in each slice of axial CT sans and the volume was obtained by the iPlan Cranial software .Orbital width was defined as the distance between dacryon and ectoconchion. The dacryon is the junction of the frontal bone, the lacrimal bone and the frontal tuber of the maxillary bone. The ectoconchion is the junction of the frontozygomatic suture and the curved surface of the orbital aditus. Orbital height was defined as the distance between the midpoints of superior and inferior orbital rims. The two midpoints were the junctions of the midnormal of the orbital width with the superior and inferior rims of the orbit. Orbital depth was defined as the distance between the lateral rim of the optic canal and the midpoint of the inferior rim .Displacement of the orbit rims was measured by comparing the mirror image of the unaffected orbit to the affected orbit. Displacement distance of superior midpoint, inferior midpoint, dacryon and ectoconchion represents the displacement of the superior, inferior, medial and lateral rims, respectively .To minimize the influence of age, microphthalmic to contralateral ratio (MCR) of the orbital variables was studied instead of absolute values. MCR was defined as the ratio of the orbital variables of microphthalmic eye to the contralateral unaffected eye, including MCR of orbital volume, MCR of orbital width, MCR of orbital height and MCR of orbital depth.To reach an unambiguous definition of the orientation of the reconstructed orbit, we used the following methods. First, on lateral position, the 3D image of the orbit was adjusted until both mandibles coincide and the Reid’s base line was horizontal. This was defined as “standard lateral position”. The Reid’s base line was the line drawn from the inferior margin of the orbit  to the superior margin of the orifice of the external acoustic meatus (Auriculare point) . After tTo measure the relative position of the cyst, we need to define the centers of the cyst and globe. The center of the cyst was defined as the point located centrally on axial, sagittal and coronal views at the same time . Center The position of the cyst was identified in three ways. First, on standard lateral position of the 3D image, by comparing the positions of the centers of cyst and globe, the cyst was defined as anterior to, posterior to or at the equator of the globe . Second,Orbital parameters were compared between microphthalmic eyes with and without orbital cysts. Statistical analysis was conducted with SPSS 19.0 for Windows . The normality test of numerical variables was conducted using the Kolmogorov-Smirnov test. Numerical variables that met the normal distribution were described as mean ± standard deviation and Student’s t test was used to compare differences between groups. Numerical variables that did not meet the normal distribution were expressed as median  and the differences between groups were tested by Mann-Whitney U test. Qualitative data were described by frequencies and proportions and tested by Chi-square test or Fisher’s exact test. Correlation analysis was expressed as Pearson Correlation Coefficients. Two-sided p<0.05 was considered as statistically significant in all analyses.Among the 20 patients with orbital cyst, 13 were males and 7 were females, age range was 0.67 to 32 years. The control group was composed of 53 males and 47 females; age range was 0.08 to 58 years. There was no statistical difference in gender (p = 0.325) or affected orbit (p = 0.742) between patients with cyst and the control group. However, significant difference was observed in ages of the two groups (p = 0.031) . All of The volume of the cysts ranged from 0.042ml to 7.551ml  ml). One patient (patient No. 16) had two cysts . The other 19 patients had only one cyst. 38.1% (8/21) of the cysts were anterior to or at the equator (anterior group). 61.9% (13/21) were posterior to the equator (posterior group). Cyst volume of anterior group was significantly larger than that of posterior group (p = 0.005). All of the cysts of anterior group were outside the muscle cone, while the majority (84.6% (11/13)) of the cysts posterior to the equator were inside the muscle cone. Among the 8 anterior cysts, 6 located infratemporally, 1 located supratemporally and 1 located infranasally. In the posterior group, 5 were right behind the optic nerve  and the rest 8 located in four different quadrants .There were significant differences between microphthalmic patients with and without orbital cyst in MCR of orbital volume (p<0.001), MCR of orbital height (p = 0.004) and MCR of orbital width (p = 0.043). For microphthalmic patients without cysts, the average volume of the affected orbit was 79.6% of the contralateral unaffected orbit, while for the patients with orbital cysts this ratio increased to 90.1%. The difference in MCR of orbital depth was not significant (p = 0.299) . MCR of The orbital rims of microphthalmic eye were displaced compared with contralateral unaffected eye in microphtahlmic patients. This diTo avoid the influence of age, age-matched controls were selected randomly from each age group at a cyst/control ratio of 1:2.The age-matched control group was composed of 17 males and 23 females; age range was 0.58 to 58 years, the meridian of age was 9.5 years. There was no statistical difference in gender (p = 0.100), age (p = 0.742) or affected orbit (p = 0.713) between patients with cyst and the age-matched control group.Significant differences were found between patients with orbital cyst and age-matched controls in MCR of orbital volume (p<0.001), MCR of orbital height (p = 0.032) and displacement of inferior orbital rim (p = 0.007). The differences in MCR of orbital width and MCR of orbital depth were not significant  . The decFor patients with orbital cyst, correlation analysis revealed that higher cyst-plus-globe volume was associated with higher MCR of orbital volume , cyst volume was also associated with MCR of orbital volume , but globe volume of the affected eye was not associated with any of the orbital parameters .We divided the patients with orbital cyst into two groups according to the cyst-plus-globe volume . The median of cyst-plus-globe volume (2.80ml) was set as cut-off point when grouping patients in this study. There were significant differences between patients with large and small cyst-plus-globe volume in MCR of orbital volume (p = 0.002), MCR of orbital height (p = 0.014), MCR of orbital width (p = 0.005), MCR of orbital depth (p = 0.002) and orbital rim displacement in inferior (p = 0.004) orbital rim . No statThe prevalence rate of microphthalmos is about 1.5~1.8 per 10,000 births,, 16 and Previous studies showed no obvious gender predilection of microphthalmos with orbital cyst., 12 In oMicrophthalmos with orbital cyst could be associated with abnormalities of different organs or systems, including cardiovascular system, central nervous system, urologic system, pulmonary, digestive tract and cleft lip/ palate., 12, 23 In literatures, microphthalmos is assessed when the axial length, adjusted for age, is below the 95th centile in children or less than 18.5mm in adults. This defCurrently, CT scanning and magnetic resonance imaging (MRI) are the most popular methods to measure orbital volume. Compared to MRI, CT scanning has several advantages. 1) It is more economical and time saving. For children, it takes only 2~3 min to do a CT scan of the orbit, while an MRI examination needs 20~30min, which is too long for small children to coordinate even under sedation. 2) CT is outstanding in evaluation of skeletal structures including the orbit. It provides excellent image quality for bony landmarks. 3) CT provides abundant data for reconstruction of 3D image of the orbit and offers accurate and reproducible measurements. However, less exposure of radiation and higher resolution of soft tissue are superiorities of MRI. In our study, great attention was paid to the protection of children from radiation. The pediatric CT parameters were carefully optimized according to the ALARA principle by reducThe position and growth pattern of the cysts could be divided into two categories: 1) anterior and out of the muscle cone, 2) posterior and inside the muscle cone. 61.9% of the cysts located posterior to the equator. The anterior cysts tended to be larger than the posterior ones. Most of the anteriorly growing cysts located infratemporarily 6/8). Large cysts growing anteriorly usually appeared as bluish orbital masses expanding the inferior eyelid, causing severe cosmetic problems and making it impossible to wear prosthesis, as described in many case reports., 6, 8, 9. Large cBecause orbital volume and other orbital variables increase with age until the end of puberty, MCRs instead of the absolute values were compared between different groups to minimize the influence of age. Moreover, the final goal of treatment was symmetry of the two orbits, not the absolute volume of one side. Among the 120 participants, statistical analysis showed very weak correlation between age and MCR of orbital volume  and no correlation between age and all the other parameters. To further avoid the influence of age, we selected an age-matched control group and got similar study results.The orbit is supposed to develop with the craniofacial skeleton from birth to teenage hood. However, the small, hypoplastic globe in microphthalmic children is inadequate to stimulate the normal growth of orbit and results in hemifacial deformity, causing severe cosmetic problems and psychological challenges in the patients. Pressure was found to be an effective stimulant in expansion of craniofacial skeleton. Orbital implants designed to stimulate orbital bone growth and socket enlargement have been proved to be effective.–30 Tse eOur previous study demonstrated that for microphthalmic children, the decrease in orbital volume resulted from decrease in the cross-sectional area of the orbital aditus , rather than the orbital depth. In the pStudies have demonstrated that orbital enlargement effect is proportional to the volume implanted. Considering that the orbital contents were composed of the cyst as well as the microphthalmic globe, we took into account the globe volume. According to our research, the cyst-plus-globe volume was positively correlated with MCR of orbital volume. Moreover, there were significant differences in MCRs of orbital volume, height, width and depth between patients with large and small cyst-plus-globe volume, but the differences were not significant between patients with small cyst-plus-globe volume and controls, which indicate only large cyst was associated with better developed orbit.Patients with orbital cyst also showed decreases in displacement of orbital rims, especially the inferior and lateral rims . This may be due to the fact that most of the cysts were attached to the inferior portion of the globe . According to our previous research, the retardation of orbital development in microphthalmic patients was mainly revealed in that of inferior and lateral rims. When an Currently, there is no consensus on the treatment strategy for microphthalmos with orbital cyst. Duke Elder describeThere are some limitations of this study. Since micropthalmos with orbital cyst was a very rare condition, the sample size of this study was small. There were not enough patients in each age group for us to analyze the influence of orbital cyst on orbital development. From this cross-sectional study, we could reach the conclusion that patients with orbital cyst had better developed orbits, but we could not interpret the cause and effect between the cyst and the development of orbit. Long-term observations on the orbital development should be carried out in the future.In conclusion, by comparing parameters including MCRs and displacement of orbital rims, we found that microphthalmic eyes with large cyst-plus-globe volume had better similarity with the contralateral eyes, comparing with microphthalmic eyes without orbital cyst or with small cyst-plus-globe volume. This might indicate the positive role of orbital cyst  in orbital development of microphthalmic patients.S1 File(XLS)Click here for additional data file.S2 File(XLSX)Click here for additional data file.S3 File(XLSX)Click here for additional data file.


Several studies have demonstrated associations of birth weight with metabolic and reproductive abnormalities in adults. The aim of this study was to investigate the birth weight in women with PCOS and its correlation with clinical and biochemical characteristics of the syndrome.We studied 288 women with PCOS according to the NIH criteria and 166 women with normal cycle and without clinical hyperandrogenism. Birth weight and anthropometric characteristics were recorded, and levels of serum androgens, SHBG, insulin and fasting glucose were measured.Birth weight data were available for 243/288 women with PCOS and age- and BMI-matched 101/166 controls. No differences were found (p> 0.05) in birth weight among women with PCOS and normal controls. Birth weight of PCOS women was negatively correlated with DHEAS levels  and positively correlated with waist circumference  and body mass index (BMI) . Birth weight of controls was negatively correlated with SHBG levels . Women from both groups were further divided in 6 categories according to birth weight . No statistically significant differences were observed in the distribution percentages between PCOS women and controls. , .Women with PCOS do not differ from controls in birth weight distribution. However, birth weight may contribute to subtypes of the syndrome that are characterized by adrenal hyperandrogenism and central obesity. Birth weight can be considered as the “reflection” of endometrial life, representing in a large scale the maternal environment. It has been hypothesized that Polycystic ovary syndrome (PCOS) may have early origins in intrauterine life . EvidencA few studies have been conducted to investigate a causative link between women’s birth weight and PCOS and results were inconclusive –8. Some The aim of this study was to investigate birth weight in women with PCOS and its possible association with clinical and biochemical characteristics of the syndrome.We studied 288 women with PCOS (25 ± 6.1 years) and 166 women (25.7 ± 6.4 years) as normal controls. Patients with PCOS were selected from the outpatient clinics of two centers . Control women were medical or dietology students or hospitals′ personnel.Research has been approved by the Institutional Review Boards of the Hellenic Red Cross Hospital and "Amalia Fleming" Hospital in Athens, Greece and all clinical investigations have been conducted according to the principles expressed in the Declaration of Helsinki. Written informed consent was obtained from all participants.We used the National Institute of Health (NIH) diagnostic criteria for PCOS patients; less than eight menses per year and a free androgen index (FAI) greater than 5 and/or clinical hyperandrogenism (presence of acne and/or hirsutism) . Other cInclusion criteria for patients and controls: 1) women who were born full-term (38 0/7 till 41 6/7 weeks of gestation), 2) no known diabetes, hypertension or dyslipidaemia or any other medical or psychiatric illness and 3) no use of any medication including oral contraceptives, anti-androgens or metformin for at least three months prior to the study.Medical and reproductive history of PCOS patients and controls was recorded and physical examination was undertaken by endocrinologists. Birth weight and gestational age were self-reported for the majority of participants, while for the rest information was obtained from medical records. Nevertheless, in a previous study, it has been shown that there is a high degree of concordance between self-reported and actual birth weight .Anthropometric measurements including weight, height and waist circumference were performed. Clinical hyperandrogenism was assessed by the presence of acne and/or hirsutism  .Morning blood samples were drawn from all participants, after an overnight fast. Plasma glucose, insulin, total testosterone (TT), Δ4-androstenedione (Δ4A), dehydroepiandrosterone sulfate (DHEA-S) and sex hormone binding globulin (SHBG) were determined in the early follicular phase of the menstrual cycle. HOMA-IR (homeostatic model assessment-insulin resistance) was calculated by using the mathematic model HOMA-IR = glucose x insulin/405 (glucose in mg/dl) for the evaluation of insulin resistance .Plasma Glucose levels were measured by an enzymatic, colorimetric method in a Cobas Integra/400/700/800 autoanalyzer (Roche Laboratory Systems).Serum insulin was measured by an immunoradiometric assay  with a sensitivity of 1 μIU/ml and intra- and interassay coefficients of variation of 2.1% and 6.5%, respectively. Serum SHBG levels were measured by an immunoradiometric assay  with a sensitivity of 0.4 nmol/l and intra- and interassay coefficients of variation of 6.1 and 8.3%, respectively. TT was measured by radioimmunoassay  with a sensitivity of 0.086 ng/ml and intra- and interassay coefficients of variation of 6% and 8.5%, respectively. Δ4A levels were measured by radioimmunoassay  with a sensitivity of 0.03 ng/ml and intra- and interassay coefficients of variation of 4.5% and 9%, respectively. DHEA-S levels were measured by radioimmunoassay  with a sensitivity of 1.1 μg/dl and intra- and interassay coefficients of variation of 5.3% and 8.1%, respectively.χ2 test with Yates Correction. Pearson’s or Spearman’s Correlation was used to explore the association between pairs of continuous variables, as appropriate. All statistical analyses were performed using the Statistical Package for Social Sciences . A p value of <0.05 was considered statistically significant.Results are presented as mean ± SD for continuous variables. Results are presented as absolute numbers or percentages for categorical variables. Differences in continuous variables between groups were tested using the independent T-Test or Mann-Whitney U test, as appropriate. Differences in categorical variables between groups were tested using The clinical and biochemical characteristics of the study participants are shown in Birth weight data were available for 243/288 PCOS women and for 101/166 controls. No differences in birth weight were found (p> 0.05) between PCOS women (3243 ± 522 gr) and normal controls (3212 ± 451 gr) .In PCOS women birth weight was negatively correlated with DHEA-S levels  , Fig. 1,The negative correlation of birth weight with DHEA-S levels in PCOS women was confirmed when we separated them in quartiles according to DHEA-S levels. Women from the highest quartile had statistically significant lower birth weight compared to the women from the lowest quartile  .Women from both groups were further divided in 6 categories according to birth weight  and percentage of women for each category was calculated. The distribution percentages in the above categories for both PCOS patients and controls are shown in In the present study we investigated the birth weight in women with PCOS and its possible correlation with clinical and biochemical characteristics of the syndrome. We did not find any differences neither in the mean birth weight nor in the distribution percentages of birth weight categories between PCOS women and normal controls.Based on Barker’s intrauterine programming theory for chronic diseases  and the Our findings are similar to the results of two large epidemiological studies. The first one, a well-designed and family-based study from the United States of America, showed that birth weight in families with PCOS defined by NIH criteria did not differ from the general population, even when it was corrected for gestational age. Furthermore, the investigators did not find any association of birth weight with reproductive and metabolic abnormalities both in women with PCOS and their relatives, males or females . The secAt variance, two recent studies with a large number of participants have shown that the presence of PCOS is not only associated with low but with increased birth weight as well. The first one was a retrospective birth cohort study of 948 singleton female babies from Australia. The results after adjusting for gestational age, suggested that two discrete fetal programming pathways, one related to high birth weight and the other to thinness at birth, are operating towards the manifestation of PCOS in adult life . The secSimilar results yielded from two other studies which investigated the link between birth weight and PCOS from another aspect i.e. by examining the birth weight of newborns of mothers with PCOS. The first one from Chile, found a significantly higher prevalence of babies small for gestational age in mothers with PCOS compared to newborns of normal women, matched for age and weight at the beginning of pregnancy . The secA possible explanation for these conflicting data could be that PCOS encompasses various phenotypic subtypes that are dictated by the parental genetic traits of the individual, the maternal contribution during fetal life and the adult environment. In large epidemiological studies, small subtypes (with low or high birth weight) probably yield no significant differences. Another explanation may be that the relation of birth weight and PCOS risk is not linear but a U-shaped, as it has been shown for insulin resistance and type 2 diabetes . If thisAn important finding of our study was that birth weight was negatively correlated with levels of DHEA-S and positively correlated with waist circumference and body mass index only in women with PCOS. To our knowledge, this association has not been found before, and it may be relevant to the fact that previous studies had not examined these parameters or included them in statistical analysis.DHEA-S, produced by the adrenal glands, is an important pro-hormone of sex steroids from in utero and throughout life. Adrenal glands are vital organs for survival. Adverse endometrial conditions, which are reflected by low birth weight, may modulate the function of such organs, by changes in blood flow redistribution or in metabolic rates, a phenomenon known as developmental plasticity. This programming of “hyper-function” of the adrenal glands, which may be crucial for fetus survival in an adverse intrauterine environment, could lead to obesity and adrenal hyperandrogenism in childhood and possibly in adult life –21. SeveApproximately 20–30% of women with PCOS have elevated levels of DHEA-S and this may be the only abnormality in circulating androgens in almost 10% of these women. Furthermore, an augmented adrenal response to exogenous ACTH stimulation –27 as weObesity mainly central, is found in almost 50% of PCOS women and constitutes a characteristic feature of PCOS that can be encountered even in normal-weight patients. It can exacerbate insulin resistance as well as hyperandrogenemia and lead to an unfavorable metabolic profile –33. ObesEpidemiological studies have documented that both low and high birth weight confer an increased risk for childhood and adult obesity –37. FetaIn control women, the above correlations did not exist. However, birth weight in these women was negatively correlated with the levels of SHBG. SHBG has an important role in the regulation of bioavailable sex steroids, throughout human life  . GeneticOne limitation of our study could be the sample size. While it is large enough for a clinical study, it is relatively small from a general epidemiological point of view. Strengths of this study is the well-defined population of Caucasian women with PCOS and controls, that were all full-term born, with a similar socio-economic status.In conclusion, in this cohort of women with PCOS birth weight was similar to age- and BMI-matched control women, probably due to a neutral net-effect of combined genetic, epigenetic and maternal factors upon birth weight. However, birth weight was associated with clinical and biochemical parameters of PCOS and may contribute to the phenotypic subtypes of the syndrome. Low birth weight was associated with adrenal hyperandrogenism, while high birth weight was associated with central obesity, both of which are features of the syndrome.


Preterm birth is a serious public health problem, as it is linked to high rates of neonatal and child morbidity and mortality. The prevalence of premature births has increased worldwide, with regional differences. The objective of this study was to analyze the trend of preterm births in the state of Paraná, Brazil, according to Macro-regional and Regional Health Offices (RHOs).This is an ecological time series study using preterm births records from the national live birth registry system of Brazil’s National Health Service - Live Birth Information System (Sinasc), for residents of the state of Paraná, Brazil, between 2000 and 2013. The preterm birth rates was calculated on a yearly basis and grouped into three-year periods  and one two-year period (2012–2013), according to gestational age and mother’s Regional Health Office of residence. The polynomial regression model was used for trend analysis.2 = 0.89), and a greater share of moderate preterm births (32 to <37 weeks), which increased from 5.8 % to 9 %. The same pattern was observed for all Macro-regional Health Offices, with highlight to the Northern Macro-Regional Office, which showed the highest average rate of prematurity and average annual growth during that period . The trend analysis of preterm birth rates according to RHO showed a growing trend for almost all RHOs – except for the 7th RHO where a declining trend was observed (−0.95 a year); and in the 20th, 21st and 22nd RHOs which remained unchanged. In the last three-year of the study period (2011–2013), no RHO showed preterm birth rates below 7.3 % or prevalence of moderate preterm birth below 9.4 %.The predominance of preterm birth rate increased from 6.8 % in 2000 to 10.5 % in 2013, with an average increase of 0.20 % per year (rThe results show an increase in preterm births with differences among Macro-regional and RHOs, which indicate the need to improve actions during the prenatal period according to the specificities of each region. It is a problem .With the decline in mortality among the 1–59 month age range, deaths during the first 28 days  came to represent a growing share. The main causes of neonatal deaths in 2010 were complications from preterm birth, as well as complications during delivery and pneumonia/septicemia . FurtherAlthough Brazil has exceed the Millennium Goals set by the United Nations, with a reduction in child mortality to 15.7 in 2015, it is still among the 10 countries with the highest rates of preterm births worldwide . In the Considering that there are regional differences in preterm birth rates  it is ju2 [This is an ecological, time series study of preterm births recorded in the Live Birth National Information System (Sinasc) for residents of the state of Paraná, Brazil, from 2000 to 2013. Paraná is considered the fifth largest economy in Brazil and is located in the country’s Southern region with an area of 199,880 km2 . It has 2 .th State Regional Health Office, because they were not yet available online at the National Health Service database (Datasus). Sinasc is a nationwide information system introduced in Brazil in 1990 with the purpose of understanding the epidemiological profile of all live births which makes possible not only the formulation of health diagnoses but also promotes surveillance and monitoring of newborns, with evaluation of health actions as well as the opportunity of population based analyzes on live births [To analyze the trend of preterm births, data from 2000 to 2012 were acquired from Sinasc, and 2013 data were provided by the 15excellent, a percentage of undeclared variables below 1 %; good, between 1 % and 2.99 %; regular, from 3 % to 6.99 %; and poor quality when the percentage of undeclared variables is equal or greater than 7 % [The period between 2000 and 2013 was defined after exploring the completion of the variables listed on Sinasc, by analyzing the rate of undeclared variables, that is, variables that were ignored or left blank. To determine the quality of the database and its potential it was used a scale which was adapted to fit the reality of the satisfactory quality of information filled into Sinasc in the state of Paraná: than 7 % . The yeaThe rates of preterm births (<37 weeks) were analyzed and calculated in relation to total live births. Preterm birth rates were calculated according to gestational age (GA): under 28 weeks (extreme prematurity); from 28 to <32 weeks (very premature); and from 32 to <37 weeks (moderate preterm birth), and mother’s place of residence  .The preterm birth rates were calculated year-to-year and also classified into gestational week intervals for the trend analysis, and grouped into three-year periods  and one two-year period (2012–2013) for being represented in graphics. Thematic maps were drawn for a first and last three-year periods (2000–2002 and 2011–2013), using greyscale, with lighter colors for lower rates and darker colors for higher rates. We opted for the use of three-year periods, due the stratification into gestational week intervals, which can lead to low number of events, likely to cause problems of statistical significance because of the occurrence of random fluctuations in the number of premature births.0 + β1x1), quadratic (y = β0 + β1x1 + β2x2), and cubic polynomial regression models (y = β0 + β1x1 + β2x2 + β3x3) were tested. Any trend whose estimated model reached a p-value <0.05 was considered significant. In order to select the best model, it was also considered the analyzes of the scatter plot, the value of the coefficient of determination (r2) and residual analysis . When all criteria were significant for more than one model and the coefficient of determination was similar, the simpler model was chosen. All analyses were carried out using SPSS software, version 20.1.The polynomial regression statistical model was used for the trend analysis in which preterm birth rates were regarded as dependent variable (y) and the years of the study the independent variable (x). The ‘year’ variable was transformed into the year-centered variable (x-2006) and the series were mitigated using a three-point moving average. The linear (y = βhttp://datasus.saude.gov.br/). All data were anonymized.The research project was approved by the Standing Committee for Ethics in Research of the Paraná State Health Secretariat/Workers Hospital . All data were obtained from public databases  were preterm births, rising from 6.8 % in 2000 to 10.5 % in 2013, with a higher share of moderate preterm births (32 to <37 weeks), which went from 5.8 % in 2000 to 9 % in 2013 . The highest average increase occurred for moderate preterm birth (0.17 % per year), with a high coefficient of determination of the model (r2 = 0.89) . The highest preterm birth rate was observed in 2012–2013 (12.0 %) at the North Macro-regional Health Office, which showed continued, expressive growth throughout the entire period. In contrast, the moderate preterm birth rate at the Center-East-South Macro-regional Health Office decreased slightly during the three-year period from 2003 to 2005 . The highest average rate of preterm births and mean annual growth was found for North Macro-regional Health Office, averaging 7.55 % of preterm births and an annual growth of 0.35 %. For its part, the Center-East-South Macro-regional Health Office showed the lowest mean prevalence for the period (6.06 %) and lower mean annual growth (0.13 %) . At the 11th RHO - Campo Mourão, there was a reduction in preterm births at the start of the period (year 2005) with a subsequent increase, whereas at the 1st RHO - Paranaguá, 20th RHO - Toledo, 21st RHO - Telêmaco Borba and 22nd RHO - Ivaiporã the premature rates remained constant  during the three-year period from 2000 to 2002, with a subsequent reduction (9.8 %) in the last three-year period .Noteworthy is the case of the 7iod Fig. . Also, tThe results evidenced that preterm birth rates are growing in Paraná State and vary among the Macro-regional and Regional Health Offices, which reiterates that there is heterogeneity in health indicators in general and in the prematurity rates among different regions, in Brazil and in other low-income countries.Preterm births lead to serious health consequences, with social and economic impact. Nevertheless, little is known about the inner distribution at the Brazilian states. This study analyzed the trend of preterm births in the state of Paraná. The results indicated that preterm birth rates rose from 6.8 % in 2000 to 10.5 % in 2013, with higher shares of moderate preterm birth (32 to <37 weeks), corroborating the results of other studies , 16.In the United States, prematurity rose from 9.5 % in 1981 to 12.3 % in 2008, leveling at between 12 % and 13 %, whereas in Europe these values went from 5 % to 9 % , 17. In Preterm birth is influenced by the degree of socioeconomic development of the population and the location of the study. Nevertheless, one of the possible explanations for the high rates of preterm births and their components lies also in the specificities of the characteristics of mother-child health for each region .th RHO of Apucarana , 15th RHO of Maringá , 17th RHO of Londrina . The HDI aims to measure the level of development by combining the variables life expectancy, income and education, and ranged from 0 (no human development) to 1 .In the state of Paraná, in the three-year period from 2011 to 2013, the highest rates of preterm births were concentrated in the RHOs with the highest socioeconomic indicators and Human Development Indexes (HDI), such as the 16A study analyzing factors related to C-section birth in the state of Paraná detected a link between C-section birth and preterm birth, and also showed an association between this type of delivery and better socioeconomic conditions . SimilarBrazil has one of the highest rates of C-sections worldwide , with a th RHO - Apucarana shows the most significant relative increase in moderate preterm birth during the period (333 %), which went from 3.7 % in the 2000–2002 three-year period to 15.9 % in 2011–2013. Thus, that RHO requires greater attention in order to detect the factors associated with this expressive increase, as even newborns between 32 and 36 weeks are immature and have significantly higher rates of morbidity and mortality compared to those born at full term [The 16ull term .Deficits in the effectiveness of preventive actions offered to expectant mothers during the prenatal period could partially explain the high percentage of prematurity, given their close link to preterm birth ; neverthth RHO - Pato Branco was the only one to show a declining trend for prematurity, with a 20.1 % from 2000–2002 to 9.8 % in 2011–2013, due especially to the decrease in the moderate preterm birth component, which fell from 18.8 % to 12.8 % during that period. Some reasons may justify this reduction: the improved Human Development Index of the city, which went from 0.71 in 2000 to 0.78 in 2010, with the expansion in the access of formal education – the aspect that grew the most in absolute terms [The 7th RHO – Campo Mourão, which showed a decrease in preterm birth rates at the start of the period, given the percentage variation of 6.05 % in 2000, 6.77 % in 2001, 7.83 % in 2002, 6.94 % in 2003 and 6.91 % in 2004, falling to 4.88 % in 2005, with a subsequent increase. With regard to the 1st RHO - Paranaguá, 20th RHO - Toledo, 21st RHO - Telêmaco Borba and 22nd RHO – Ivaiporã, there was no statistically significant trend due to the fluctuation in prematurity rates in the study period.Highlight must also be given to the 11The object of this study was not to identify the factors associated with preterm births in the state of Paraná; nevertheless, it is important to emphasize that improved access to healthcare, with the expansion in ESF coverage statewide, was not capable of reducing the occurrence of preterm births.In addition to elective C-sections, induced labors and unsatisfactory quality of prenatal care , 25, it th RHO - Maringá, 17th RHO - Londrina and 18th RHO - Cornélio Procópio, while the rates of very premature births doubled at the 15th RHO - Maringá and 2nd RHO – Metropolitan. It is important to remember that extremely premature and very premature births represent a high risk for severe morbidities, hospital re-admissions, permanent disabilities and mortality, possibly leading to severe economic, social and emotional consequences for the family and society [Another noteworthy result is that although extremely premature (<28 weeks) and very premature (28 to <32 weeks) births have showed low average annual growth rates, their distribution in the thematic maps revealed that the prevalence of extremely premature births doubled at the 15 society , 27.Some limitations of this study are noteworthy, such as those related to information obtained from secondary databases, especially the possibility of differences in the reliability of data among the different regions. Thus, whenever long time periods are analyzed and different regions are compared (as in this study), any trend in preterm birth rates may be hidden by incorrect entry of gestational age. Also any increase trend could be partly related to gradual improvement in the quality of data entry in declarations of live births.In addition, the accuracy of the information about gestational age that is registered in the Sinasc database may have discrepancies among the State regions and during the long period of time. Until 2009, the Sinasc database system grouped the gestational age variable in an age strata, only after this year it has been recorded according to exact number of gestational weeks. This made it impossible to analyze borderline prematurity (34 to <37 weeks) and likely induced classification errors, with a tendency towards recorded gestational age ranges closer to full-term births.Despite these limitations, the data available in Sinasc national system and other databases that have similar health information in Brazil make possible to know and monitor the health situation of the population and identify differences among cities, regions, states and macro-regions. These results allow the evaluation of healthcare policies and interventions, as well as possible inequalities in life, education and healthcare conditions.Despite the policies implemented to prevent the preterm birth, our results showed increase of this event in the Southern of Brazil, with differences between Macro-regional and Regional Health Offices. These results confirm what is observed in other countries. However, further studies must be carried out in order to understand the role of socioeconomic and demographic factors, healthcare and organization of health services in the occurrence of preterm births, as well as how risk factors are distributed in different regions and whether the variations in the preterm birth rates can be attributed to local characteristics.The increase of preterm birth rates shows that measures to prevent this event need to be improved, especially with the moderate preterm births. Births which occurred from 32 to less than 37 gestational weeks are usually related with caesareans, that has extremely high rates in Brazil and in another countries with the same medicalized model of assistance. Therefore, is also important to monitor the rates of unnecessary caesarean and direct the public polices to the regions with the highest rates.It is worth mentioning the potentiality of use of secondary data provided by the Sinasc. The analysis of variables available in this health system allows to know and monitor the population's health situation, identify differences between the regions, showing results of health care provided in each locality, and possible inequalities in quality of life, education and assistance for the maternal and child health.http://datasus.saude.gov.br/). All data were anonymized.The research project was approved by the Standing Committee for Ethics in Research of the Paraná State Health Secretariat/Workers Hospital . All data were obtained from public databases (Not applicable.http://datasus.saude.gov.br/).The dataset supporting the conclusions of this article is available from public databases in the DATASUS repository (


The aim of the present study was to present a novel non-local mean (NLM) method to denoise diffusion tensor imaging (DTI) data in the tensor space. Compared with the original NLM method, which uses intensity similarity to weigh the voxel, the proposed method weighs the voxel using tensor similarity measures in the diffusion tensor space. Euclidean distance with rotational invariance, and Riemannian distance and Log-Euclidean distance with affine invariance were implemented to compare the geometric and orientation features of the diffusion tensor comprehensively. The accuracy and efficacy of the proposed novel NLM method using these three similarity measures in DTI space, along with unbiased novel NLM in diffusion-weighted image space, were compared quantitatively and qualitatively in the present study. The Bro smoothing , was proin vivo DTI data were assessed.The present study introduces the basic principle of this smoothing method, as well as the implementation details. Performance of the algorithm was evaluated on the basis of three similarity measures, namely, Log-Euclidean distance (LED) , Riemannet al  is an ED of the gray level between pixels p and q, h controls the rate of decay of the exponential function and Z is a normalizing factor calculated as follows ::16):V is the tensor matrix and λi is the ith eigenvalue of the matrix, V(p)−1V(q). Compared with the Euclidean metric, which admits non-positive tensors and exhibits a swelling effect, the affine invariant metric coincides with the Fisher information metric ::V is theTrace denotes the sum of the eigenvalues.where According to the original NLM method , the grawhere the weight can be calculated using equations 6in vivo conditions, synthetic tensors in the curves were constructed to have a trace of 2.1×10−5 cm2/sec and an FA of 0.8. These diffusion parameters were similar to those in normal brain parenchyma. Diffusion-weighted imaging was simulated along 32 non-collinear directions with a b value of 1,000 sec/mm2, and the diffusion-weighted data were corrupted with 5% Rician noise as follows  and mean deviation of the FA prior to and following denoising. The first assessment demonstrated the improvement of the visual effect with the denoising methods. The second assessment measured the capability of these denoising methods to restore the tensor orientation, while the last assessment gauged the extent to which the tensor directionality was restored with the denoising methods.Finally, the effects of denoising were more rigorously examined with permutation tests, which compared the influences of different denoising methods on the uncertainties of the DTI data. Experimental procedures were conducted as follows. Firstly, the DWI data were computed from the denoised and uncorrupted DTI data. Secondly, residuals were obtained by calculating the differences between the DWIs from the denoised DTI data and those from the uncorrupted DTI data. The residuals were then randomly permuted and added to the uncorrupted DWI data. This process was repeated 1,000 times to produce 1,000 new DWI datasets, from which DTI data were derived. The 1,000 samples of each of the six independent tensor components were fitted with normal distributions, and the mean and 95% confidence intervals for each component were calculated. Finally, an ellipsoid was constructed corresponding to the mean DTI data, which was sandwiched by ellipsoids corresponding to the upper and lower bounds of the confidence interval.h controls the decay of the exponential function, which impacts the degree of filtering. Typically, the value of h is selected on the basis of the variance of noise  and an eight-element SENSE coil. A single shot, echo-planar pulsed gradient spin-echo imaging sequence was used, and diffusion weighting was performed along 32 non-collinear directions with a b value of 1,000 sec/mm2. A total of 64 contiguous, 2-mm-thick slices with a matrix size of 128×128 were acquired from a field of view of 256×256 mm2, yielding an in-plane pixel size of 2×2 mm2. Three repeated scans were obtained from each subject, which were motion and distortion corrected and then averaged using the Philips diffusion registration PRIDE tool . Diffusion tensors were estimated from the averaged DWI data using a linear least-square fitting procedure (To assess the performance of the proposed algorithm on in vivo dataset and the parameter settings were the same as for the synthetic experiments. Denoising results were evaluated visually and quantitatively, on the basis of the tensor geometry, the mean angular deviation of PD and the mean deviation of FA, as before.The same four methods were utilized with the Results of the proposed DTI-NLM method are demonstrated in Among the different weighting schemes using NLM, the RD and LED methods  yielded Quantitative evaluations of the smoothing effects are shown in Changes in the tensor uncertainties following denoising are shown in in vivo human DTI dataset, which demonstrated marked artifacts resulting from noise. in vivo data were similar to those on the simulated data.To assess the denoising effects in the tensor structure, the same section presented in The computational efficiency of the proposed NLM-DTI algorithm with the three different weighting schemes and UNLM in DWI space is demonstrated in In the present study, a novel DTI denoising method in the tensor space is proposed. Following the widely recognized NLM method, DTI data were denoised by averaging the tensors in a specific neighborhood according to their similarity to the tensor in the center. Under the proposed NLM framework, three weighting schemes corresponding to different features extracted from the tensor, as well as GF in DTI space and UNLM denoising in DWI space, were compared quantitatively and qualitatively with synthetic and real human brain DTI datasets.The results of these experiments demonstrated that the proposed method is able to suppress noise and preserve explicit boundaries in DTI data, with the extent varying depending on the similarity weighting scheme used. Specifically, the denoising performance relied primarily on the accuracy of tensor comparison. When the tensor was compared on the basis of a greater number of structural features, as implemented in RD and LED, the proposed NLM method achieved the best effect. By contrast, when the comparison used fewer structural features, as implemented in ED, the denoising effect was markedly decreased.b values. In addition, due to the nonlinearity between the DWI and DTI spaces, optimal denoising results in the DWI space are not necessarily optimal in the DTI space. Therefore, denoising DTI directly is able to achieve the most intuitive restoration results and convenience for subsequent applications.As previously discussed, DTI denoising may be broadly divided into two categories: Denoising in the DWI space and denoising in the DTI space. Compared with DWI denoising, DTI denoising is much more convenient to implement and is able to achieve similar denoising results for orientation and structure information. Unlike DTI denoising, which involves only six independent parameters, DWI data normally entail a much greater number of directions and the denoising procedure has to be duplicated for all the directions. As the application of HARDI increases , the comin vivo human DTI data. With the application of an affine invariant matrix, which encapsulates the comprehensive information of the diffusion tensor, NLM-DTI denoising is able to achieve comparative restoration results to resemble the denoising method in DWI space. In addition, the time efficiency demonstrates the practical use for this method as a routine denoising algorithm.Finally, it should be considered that the proposed method does not come without limitations, despite its great potential. Although denoising in DTI space is timesaving when compared with DWI denoising, the NLM method itself is generally computationally intensive. This can be resolved by improving the NLM framework  or by im


To evaluate in a national standardised setting whether the performance of ultrasound dating during the first rather than the second trimester of pregnancy had consequences regarding the definition of pre- and post-term birth rates.st trimester), BPD (2nd trimester), and head circumference and compared mean and median durations, the mean differences, the systematic discrepancies, and the percentages of pre-term and post-term pregnancies in relation to each method. The primary outcomes were post-term and pre-term birth rates defined by different dating methods.A cohort study of 8,551 singleton pregnancies with spontaneous delivery was performed from 2006 to 2012 at Copenhagen University Hospital, Holbæk, Denmark. We determined the duration of pregnancy calculated by last menstrual period, crown rump length (CRL), biparietal diameter (1nd trimester) is used. Calibration for these discrepancies resulted in a lower post-term birth rate, from 3.1–1.4%, when first compared to second trimester dating was used.The change from use of second to first trimester measurements for dating was associated with a significant increase in the rate of post-term deliveries from 2.1–2.9% and a significant decrease in the rate of pre-term deliveries from 5.4–4.6% caused by systematic discrepancies. Thereby 25.1% would pass 41 weeks when GA is defined by CRL and 17.3% when BPD (2Systematic discrepancies were identified when biometric formulas were used to determine duration of pregnancy. This should be corrected in clinical practice to avoid an overestimation of post-term birth and unnecessary inductions when first trimester formulas are used. The accurate dating of the duration of pregnancy is of importance in regard to prenatal care since several clinical decisions are based on gestational age (GA). In most industrialised countries, ultrasound (US) is accepted as the method of choice, yet in the literature, 7- to 14-day discrepancies are described. The discrepancy includes biological variation in the duration of pregnancy and methodological errors.  Such a dUS dating has been investigated in several studies, and biparietal diameter (BPD) in the second trimester has been found to be superior to last menstrual period (LMP). , 3, 4 ThIn Denmark, an effort to standardise pregnancy dating has been made through the establishment of a national guideline. All pregnant women are offered a first trimester US scan, in which CRL and nuchal translucency (NT) are measured and GA is determined. This examination is a part of a free-of-charge prenatal programme combining age, biochemistry, and US screening for chromosomal abnormalities in the first trimester and a second trimester screening for malformations. The compliance rate of these two screenings has increased rapidly over the past few years, to the current 94%.  In 2007,The aim of the present study was to compare dating during the first and second trimester and determine their consequences for rates of pre-term and post-term births.The results of this cohort study were reported following the STROBE recommendation. The study was approved by the Danish Data Protection Agency and the regional Research Ethics board (reg. no. SJ-HO-01 and SJ-335). Written informed consent was not required and therefore not obtained. However, patients record and the data were anonymized before analysis.www.astraia.com), which comprised all US examinations performed on pregnant women during the study period and includes the mothers’ medical history, LMP, parity, pre-pregnancy maternal weight and height, and smoking and alcohol habits. The data were linked by the personal civil registration number with data from the Danish Medical Birth Registry, which contains data on the date of birth, birth weight and length, infant sex, number of infants born, intrauterine foetal death, pregnancy loss and complications during birth. All births in Denmark are registered in the Danish Medical Birth Registry and at the time of data extraction the register were administered by the Danish National Board of Health. [Data were collected prospectively from 2006 to 2012 at Copenhagen University Hospital, Holbæk, in Denmark. The population was an unselected population of pregnant women, and data were retrieved from the Astraia database  Inclusion criteria were a registered CRL and the later birth of a live-born child. The following pregnancies were excluded: a CRL registered < 45 mm and > 84 mm , multiple pregnancies (n = 530), induction of labour , elective caesarean section (CS) or acute CS before labour , and stillbirths (n = 21). A total of 8,551 pregnancies were included in the study. , BPD2 between 31–55 mm (GA: 15–22), and HC 110–200 mm (GA: 15–22). Sub-analyses were performed in which unregistered LMP (487), BPD1 (132), BPD2 (367), and HC (822) were excluded when analyses included one of these parameters. Absence of this information meant that foetometry was not performed during the given interval, or that one of the parameters was not registered.BPD and HC measurements were categorized in the following intervals: BPDst day of the LMP, the CRL (Robinson and Fleming formula) [The GA was calculated using the 1formula) , and theformula) These formulas were chosen since they are in accordance with the Danish national guidelines. GA at birth was computed from GA defined at the time of the US examination (depending on method used) plus the number of days from the US scan to the date of birth.The biometric measurements were done in accordance with the instructions provided by the authors of the above-mentioned formulas. The ultrasound examinations were all performed by midwives and doctors trained and certificated to perform NT measurements by the Fetal Medical Foundation, London, UK.A delivery at ≤ 258 completed days was considered pre-term, a delivery between 259–293 completed days to be term, and a delivery at ≥ 294 completed days to be post-term.1 in the first and BPD2 and HC in the second trimesters were analysed by comparing GAs ) at spontaneous delivery, defined by the date of birth and a spontaneous onset of birth. The methods were tested for mean differences compared to different methods to take account for the normal variation of the pregnancy length. The mean difference is the methodological differences and SD is the random insecurity in-between the methods.CRL and BPDet al. [To evaluate systematic errors in the prediction models, the errors were calculated by subtracting the predicted date from the median duration of the pregnancy as suggested by Okland et al.  The medi1, BPD2, HC, and LMP.The percentages of deliveries with GAs of ≤ 247, ≤ 258, ≥ 287, ≥ 290, and ≥ 294 on the day of birth were compared between the methods CRL, BPDp value of < 0.05 was considered statistically significant.McNemar’s test was used in analysing paired data. All statistical analyses were performed using SPSS, version 21. All tests were two-tailed and a There were up to three days differences in median gestational age depending on the method employed. The SD was smallest for first trimester dating .1, and +1 day for BPD2 and HC.The systematic discrepancies (defined as median difference by subtracting the medians in nd trimester) defined most pregnancies within the interval of term (93.6%), BPD1, (92.8%), BPD2 (92.2%), CRL (92.2%) and LMP (86.0%). LMP defined most pregnancies as post-term (9.9%) and HC fewest (0.9%), both significantly different compared to CRL.2. However, there was still a significant difference in post-term rate between CRL and BPD2, with an increase in pregnancies defined as post-term by BPD2 (3.1%) (1 (93.5%), BPD2 (92.0%), HC (93.4%)), but different for LMP (88.1%).Calibration of the prediction models showed that the distributions of pregnancies defined as pre-term, term, and post-term were more even, especially between CRL and BPD2 (3.1%) . The disIn a large cohort of pregnant women, the use of first trimester dating resulted in significantly more post-term pregnancies compared to second trimester dating.2 (2.1%). Though, when the formulas in our material were calibrated to a duration of pregnancy of 280 days, the opposite was found. [This is of special interest due to the on-going discussion on whether induction of labour should be performed at 41 + 0 instead of 42 + 0 weeks in low risk pregnancies. We found a significant decrease in post-term birth rate when comparing LMP to BPD dating in the second trimester, which is in agreement with previous studies.  However,s found. 2 is used, and 31% when LMP is used.Our results showed that if the time of induction was changed from 42 + 0 to 41 + 0, the difference between the percentages of pregnancies reaching 41+ 0 defined by first versus second trimester calculation increased to 7.8%. This means that 25.1% of all pregnancies would pass 41 weeks when GA is defined by CRL, 17.3% when BPDWhen calibrating the formulas by assuming that the median duration of pregnancy should be 280 days, the post-term birth rate decreased using first trimester biometrics. This is not standard in our national setting, but should be considered to diminish these differences. Alternatively, other and more precise formulas could be used as population-based prediction methods. , 15 The et al. argued that first trimester US would theoretically decrease post-term rates compared to second trimester US; however, the study did not include second trimester measurements. [et al. found a lower post-term birth rate in first trimester compared to second trimester measurement, though the difference was not significant. [The current literature on this issue is relatively sparse and other studies have suggested that dating in first trimester with CRL would decrease post-term birth rate further, but this association has not been substantiated. Taipale rements.  In a ranrements.  Saltvendificant. 1 showed identical results regarding the duration of pregnancy (median and SD) and a mean difference of 0.38 (SD: 2.04) days, saying first trimester dating is more precise and with the smallest variation. In our regimen, we measure BPD routinely in the first trimester in order to detect central nervous defects. Therefore it may be easier for inexperienced personnel to measure BPD instead of CRL.There is a need to establish standardised clinical guidelines for surveillance and intervention during pregnancy. Worldwide, there are different weeks for schedules of US, formulas, and methods used for dating. In the present study, first trimester measurements CRL and BPDet al. have published two formulas [et al. we found that GAs were skewed to the right in both the first and second trimesters when using the second formula from 1997, giving a median in BPD1 of 284 and 285 in BPD2 (data not shown). [The dating methods are based on the assumption that all foetuses are of similar size at a given GA during the first half of the pregnancy, yet several factors, e.g. gender, parity, maternal age, smoking habits, etc., may influence dating. , 17, 18 formulas , 19, whi shown). , 19 HoweThe strength of the present study is that we studied a large unselected population in a standardised programme with a large number of first trimester measurements, thereby minimising potential selection bias. Exclusion of post-term induction is a limitation to the study, causing an artificially low post-term rate. By using an older cohort this selection bias was minimized, since induction-rate in general is increasing. In our material the mean induction-rate was 17.1%  comparedIn conclusion, we found an increase of the post-term rate in dating performed in first trimester compared to second trimester, caused by systematic discrepancies. When the formulas were calibrated, the post-term birth rates in the measurements performed in first trimester decreased. It is important to evaluate and correct the formulas used for dating to avoid misclassification of pre-term and post-term pregnancies and thus avoid unnecessary inductions due to post-term pregnancy.S1 Table(DOCX)Click here for additional data file.