The WHO ACTION-I trial demonstrated that dexamethasone significantly reduced neonatal mortality when administered to women at risk of early preterm birth in low-resource countries. We conducted a secondary analysis to determine how these benefits can be optimised, by evaluating the effect of dexamethasone compared to placebo on newborn mortality and severe respiratory distress outcomes at different administration-to-birth intervals, and identifying the interval with the greatest benefits.The WHO ACTION-I trial was a multi-country, individually-randomised, parallel-group, double-blind, placebo-controlled trial. It was conducted in 29 hospitals across Bangladesh, India, Kenya, Nigeria, and Pakistan. Women with a viable singleton or multiple pregnancy who presented to participating hospitals at a gestational age of 26 weeks 0 days–33 weeks 6 days and who were at risk of imminent preterm birth were eligible. In this secondary analysis, 2638 women and their newborns treated with single course of dexamethasone or placebo were analysed. Multivariate logistic regression was used to assess the effect of dexamethasone versus placebo on neonatal death, stillbirth or neonatal death, and severe respiratory distress at 24 h and at 168 h, by administration-to-birth interval (from 0 through 28 days), adjusting for gestational age at first dose. We used relative risks to identify the administration-to-birth interval with the greatest benefits of dexamethasone compared to placebo on the newborn outcomes.Between 24 December 2017 and 21 November 2019, 2852 women and their 3070 babies were enrolled in the WHO ACTION-I trial; 1332 women (1464 babies) in the dexamethasone group and 1306 women (1440 babies) in the placebo group were included in this secondary analysis. Neonatal mortality risk was lower with increasing time between initiating dexamethasone and birth, achieving peak mortality reduction by days 13 and 14 and then diminishing as the interval approached 28 days, regardless of gestational age at administration. For other outcomes, the overall pattern of risk reduction extending into the second week was consistent with that of neonatal death.In women at risk of preterm birth prior to 34 weeks’ gestation, the neonatal benefits of antenatal dexamethasone appear to increase with longer administration-to-birth intervals than previously thought. This knowledge can support clinical assessment and estimation of the risks of adverse preterm newborn outcomes at the time of birth, and the potential benefits of antenatal dexamethasone treatment for a known administration-to-birth interval.Bill and Melinda Gates Foundation; World Health Organization. We conducted a systematic search of PubMed, Embase, CINAHL, Cochrane Library and Global Index Medicus (using search terms and synonyms for “pregnancy” and “corticosteroid” and “duration”), with no date or language restrictions to identify randomised and non-randomised studies where newborn outcomes were reported for different antenatal corticosteroid administration-to-birth intervals. The review identified 59 studies , largely from high-income country settings, and there was considerable heterogeneity in study populations, administration-to-birth intervals used, and conclusions. It was not possible to identify an optimal interval from existing evidence.To our knowledge, this is the first analysis of an antenatal corticosteroid efficacy trial that has identified further reduction in the relative risk of neonatal mortality with longer dexamethasone administration-to-birth intervals. The relative risk of severe neonatal respiratory distress at 24 h and at 168 h showed similar patterns although not as remarkable as that of neonatal death. This analysis goes beyond previous studies by using administration-to-birth interval as a continuous  variable, and using multivariate modelling to account for the confounding effects of gestational age and other important maternal and newborn factors. We developed a practical tool for potential clinical application of these findings.The neonatal mortality reduction benefits from antenatal dexamethasone for women at risk of early preterm birth might be substantially higher with longer administration-to-birth intervals than generally thought, regardless of gestational age at the time of initiating treatment. Our findings question the scientific basis of clinical recommendations that are based upon the premise that ACS are not effective and even harmful after 7 days following initiation of treatment. The new knowledge generated from this study can be used to support clinical assessment and estimation of the risks of adverse preterm newborn outcomes at the time of birth, as well as the potential benefits of ACS treatment for a known administration-to-birth interval (such as in the context of provider-initiated preterm birth).Antenatal CorticosTeroids for Improving Outcomes in preterm Newborn)-I trial, which was conducted in hospitals across five low-resource countries, provide reassurance on the clinical benefits of ACS when used for women at risk of early preterm birth. The trial reported a substantial reduction in the risks of neonatal mortality alone and stillbirth or neonatal mortality, without increasing the risk of maternal or neonatal infection,Antenatal administration of corticosteroids to women at risk of preterm birth has been shown to improve outcomes for preterm newborns. The 2020 update of the Cochrane review on efficacy of antenatal corticosteroids (ACS) demonstrated that a single course of dexamethasone or betamethasone prior to 34 weeks’ gestation confers significant reductions in neonatal mortality, and severe morbidities, without causing maternal or newborn harms.Now that the benefits of ACS administration in low-resource countries have been established, there is a need to determine how these benefits can be optimised in clinical practice. One important consideration is the potential relationship between the time interval from administration of ACS to birth and how the interval might affect the neonatal benefits of ACS. In contemporary obstetric practice, it is generally accepted that there is a minimum and a maximum time of in-utero “fetal exposure” to ACS to achieve clinical benefits and avoid harms. While this notion is reasonable from a scientific standpoint , what constitutes the best  time interval to maximise clinical benefits of ACS remains unclear. Several organisations, including WHO, recommend that administration of ACS should target eligible women with a high likelihood of preterm birth within 7 days of starting treatment, but the quality of the underlying evidence is low.,Previous studies exploring the association between time from ACS administration-to-birth and newborn health outcomes have generally been inconsistent in their findings. For example, a study in Canada by Melamed et al.,,The design of the WHO ACTION-I trial has been described previously in the study protocol and primary findings.Women with a viable singleton or multiple pregnancy who presented to participating hospitals at a gestational age of 26 weeks 0 days–33 weeks 6 days and who were identified as being at risk of imminent preterm birth were eligible for inclusion and invited to participate. Imminent preterm birth was defined as planned or expected birth in the next 48 h (either provider-initiated or after preterm prelabour rupture of membranes or spontaneous labour). To be eligible, a woman's gestational age had to be informed by an obstetric ultrasound. Women were not eligible if they had clinical signs of severe infection, suspicion, or evidence of clinical chorioamnionitis, major congenital fetal anomalies, concurrent or recent use of systemic steroids, were participating in another maternal or newborn health trial, or had a contraindication to steroid use. Women were screened at the time of presentation to hospital, and randomised women and their babies were followed up until day 28 after birth, at the hospital or at home. Written informed consent was obtained from every woman prior to randomisation.For this secondary analysis, we included randomised women (and their newborns) who received a single course of dexamethasone or placebo, had complete data for trial primary outcomes, and who had complete data on time of dexamethasone or placebo administration and time of birth. Women who received a repeat course (and their newborns) were excluded. Hence, the population for this analysis comprised 92.5% of women and 90.2% of babies who participated in the WHO ACTION-I trial.In the ACTION-I Trial, site-stratified individual randomization (1:1 ratio) with balanced permuted blocks of 10 were used. Eligible, consenting women were randomly assigned to a course of IM injections of either 6 mg dexamethasone or identical placebo  administered every 12 h, to a maximum of four doses, or until hospital discharge or birth, whichever was earlier. Women were eligible for a repeat course if they had not given birth after 7 completed days but still met the eligibility criteria .The ACTION-I trial primary outcomes were neonatal death until 28 completed days of life; stillbirth or neonatal death (any baby death); and a composite outcome for possible maternal bacterial infection . Secondary outcomes included maternal and newborn mortality and morbidity outcomes, and process of care outcomes.For this secondary analysis, the outcomes of interest were neonatal death, stillbirth or neonatal death , severe neonatal respiratory distress within 24 h, and severe neonatal respiratory distress within the first week of birth (168 h). In the ACTION-I trial, severe neonatal respiratory distress was clinically defined based on the composite of fast breathing (respiratory rate ≥70 breaths per minute); and at least one of marked nasal flaring during inspiration, expiratory grunting audible with naked ear, or severe chest in-drawing; and SpO2 less than 90% or use of supplemental oxygen.The exposure variables of interest for this analysis were the allocated treatments (dexamethasone and placebo). We aimed to assess how selected newborn outcomes varied by administration-to-birth interval, which was defined as the time from administration of the first dose of dexamethasone or placebo until birth. This interval was considered using both categorical and continuous variable approaches. For each outcome, we explored the administration-to-birth time interval which was associated with the greatest benefit of dexamethasone compared to placebo.a priori might be concealed, and it does not take full advantage of having continuous data for the administration-to-birth interval. It also does not take into account the impact of gestational age  on newborn health outcomes – it was evident from this analysis as the relative risks varied considerably depending on the gestational age at time of first dose.We used two different statistical approaches. First, we conducted subgroup analyses stratifying the study population by pre-defined categories of administration-to-birth intervals, and calculated the relative risks of dexamethasone compared to placebo for each newborn outcome and for each interval category. This approach had two important limitations: firstly, the categories of administration-to-birth intervals were pre-defined based on clinical practice considerations and boundaries that were set from previous research. As a result of these limitations, interactions with intervals other than those categories defined To address these issues, we used a second statistical approach. We created a prediction model based on women receiving dexamethasone or placebo for each of the newborn outcomes of interest, taking into account the administration-to-birth interval as continuous variable, as well as gestational age at first administration. Then, from the predicted risks for the dexamethasone and placebo arms, we estimated the relative risks (RRs) for each outcome by gestational age at first dose to identify the interval with the greatest benefit. We used multivariate logistic regression to create the prediction models for each outcome. First, we fitted a full quadratic polynomial model for each trial arm on time from administration-to-birth and gestational age at first dose , plus covariates. The following covariates were ultimately included in the model: gestational age at first dose, whether the preterm birth was spontaneously- or provider-initiated, mode of birth, maternal age, singleton or multiple birth, nulliparity, history of preterm birth, having at least one obstetrical condition present, and whether tocolytics and magnesium sulfate for fetal neuroprotection had been administered. These covariates were all measured at baseline, except for mode of birth.We used the Lasso strategy for selecting variablesUsing the reduced model, we estimated for each gestational age at first dose the effects of dexamethasone by different administration-to-birth intervals in days up to 28 days from the start of treatment, in order to identify the time interval with the greatest benefits of dexamethasone compared to placebo. Effects of dexamethasone were expressed as relative risks . To estimate the relative risks with their 95% confidence intervals, we used the Poisson distribution with the log link in the logistic model and then estimated linear combinations corresponding to the relative risks for different values of gestational age at first dose and administration-to-birth intervals. The same analysis was conducted with administration-to-birth intervals in hours up to 24 h from the start of treatment to further elucidate the relative risks of neonatal death at shorter intervals. For each gestational age at first dose, we plotted the relative risks by administration-to-birth intervals, and identified the optimal interval as the time points where greatest impact of dexamethasone (in terms of risk reduction compared to placebo) were observed.To aid interpretation and develop a tool for potential clinical application of these findings in our study context, we used profiler software . The review identified 59 studies , 39 of which were conducted in high-income country settings. One of the 11 trials was the WHO ACTION-I trial primary analysis, and none of the other 10 trials evaluated ACS administration-to-birth interval as a continuous measure, nor adjusted for the confounding effect of gestational age at time of administration. Findings in the current analysis have not been reported previously in efficacy trials. Three trials reported on neonatal mortality for different categories of ACS administration-to-birth interval - all three were small (ranging from 73 to 208 newborns) and reported no differences in neonatal death across intervals (15–17). Seven trials (with 80–315 newborns) reported that odds of RDS were not different for any ACS administration-to-birth interval . However, one trial (696 newborns) found reduced odds of RDS at 1–7 days and >7 days with ACS compared to no ACS treatment (23). One other trial (73 newborns) found reduced odds of RDS at ≥8 days compared to no ACS treatment (24).,,,Large observational studies from Canada and Australia, as well as multi-country studies from other high-resource settings, have explored the relationship between the ACS administration-to-birth interval and newborn outcomes.The strengths of this study include the use of data from a multi-country, randomised, placebo-controlled, double-blind trial, in which the administration-to-birth interval was prospectively reported with high precision, and newborn health outcomes were measured robustly and systematically, and there were high event rates for newborn primary outcomes. This secondary analysis included 92% of women and 95% of babies who participated in the original trial. This secondary analysis was anticipated during the design phase of the trial and thus allowed for a proactive collection of accurate data on time of treatment and time of birth. Multiple analytical approaches were used to explore associations between administration-to-birth interval and newborn outcomes. The robustness of the data set for neonatal mortality outcome allowed for construction of a tool that could potentially be applied in the clinical context of the countries where WHO ACTION-I trial was conducted. Nonetheless, we acknowledge several limitations, including that this was a secondary analysis restricted to trial participants who received a single course of dexamethasone or placebo. The analysis relied on a post-randomisation exposure variable  and is therefore prone to confounders which were not systematically captured. The relationships between fetal maturation, time and dexamethasone are complex, and residual confounding, or other unknown factors, may be playing a role. We have a limitation in drawing reliable inferences for the two clinically assessed severe respiratory distress outcomes, as the risk and relative risk estimates for these outcomes have low precision. This might be in part explained by the lower prevalence recorded for these secondary outcomes in the trial population, probably as a result of some degree of under-recognition of newborn respiratory complications based on clinical assessments alone (3). This is in contrast to the prevalence of neonatal mortality, which was the primary outcome upon which the trial sample size estimation was based. There were also fewer observations with longer administration-to-birth intervals, leading to greater uncertainty . A number of other factors related to the analysis population might affect generalizability of these findings, such as the enrolment of women at risk of imminent preterm birth from 26 weeks 0 days–33 weeks 6 days only; that two-thirds of women had a gestational age estimate based on third trimester ultrasound; and that the ACS regimen was IM dexamethasone phosphate. Further validation of these findings in other, high-quality datasets would be useful.Our study confirms that the window of benefit in terms of neonatal mortality reduction following dexamethasone administration is time bound, but the inflection point and subsequent diminution do not occur until about two weeks following the initiation of treatment. In view of the challenges of accurately predicting the timing of spontaneous preterm birth, this longer optimal administration-to-birth interval provides reassurance regarding the benefits of ACS in situations where anticipated birth occurs after the conventional 7 days’ window. As the mortality risk among infants who were exposed to a single course of dexamethasone is lower for the larger part of 28 days following treatment, administration of a repeat (or “rescue”) course during this period to optimise the administration-to-birth interval might be unnecessary. However, this warrants further investigation, as trials of repeat ACS regimens have thus far used betamethasone only (27).From a clinical standpoint, clinicians may consider prolonging the time of birth with tocolytics where it is deemed clinically safe, not just for ensuring that all prescribed doses of ACS are administered, but also to optimise the administration-to-birth interval, while continually weighing the balance of benefits of increasing interval against risk of fetal exposure to an adverse intrauterine environment. The tool developed from WHO ACTION-I data could be applied in hospitals in low-resource countries with similar characteristics as in the trial to estimate risks of dexamethasone treatment versus no treatment, in situations where the administration-to-birth interval can be pre-determined with a high degree of certainty, such as in provider-initiated preterm birth. The tool could also be used for support clinical assessment and estimation of the risks of adverse preterm newborn outcomes at the time of birth when all variables, including the administration-to-birth interval, are known.Our findings call for a review of the 2015 WHO recommendation that suggests that antenatal corticosteroid should only be applied when preterm birth is considered “imminent” within 7 days of starting treatment, which was based on subgroup analysis that is no longer supported in the current updates of Cochrane review underpinning the recommendation.Further investigation is required on the overarching question of what administration-to-birth interval confers the best clinical outcomes for preterm newborns. We acknowledge that it is methodologically and ethically impossible to randomise women to different administration-to-birth intervals. Therefore, combining existing WHO ACTION-I trial data with similar high-quality data sets on administration-to-birth could be the starting point to strengthen available evidence. Any future studies on this question, including observational studies, should analyse prospectively documented administration-to-birth interval as a continuous measure, and avoid arbitrary cut-offs that tend to hide the nuances in the risks of adverse outcomes within administration-to-birth interval categories. Future research could also benefit from individual patient data meta-analysis to produce more generalisable information and understand whether the observed patterns differ by population subgroups of women at risk of preterm birth. The application of machine learning to large observational and randomised trial data sets to improve prediction of the timing of preterm birth, and individualise risk estimation by time of administration-to-birth interval is long overdue.In conclusion, this secondary analysis of a multi-country randomised trial indicates that the neonatal mortality reduction benefits from antenatal dexamethasone for women at risk of preterm birth prior to 34 weeks’ gestation may be substantially higher with longer administration-to-birth intervals than previously thought. Risk reduction for important neonatal outcomes tend to continue to increase into the latter part of second week from the start of dexamethasone administration, regardless of gestational age at time of administration.The concept for this secondary analysis was conceived by O.T.O. and J.P.V., which was developed with G.P., J.C., and A.D.C. Statistical analysis was performed by G.P. and J.C., with findings presented and reviewed by all authors. J.P.V. and O.T.O. wrote the first drafts of the article, with input from F.A., G.P., R.B., S.P.N.R. and A.D.C., which was then reviewed and revised by all named authors. G.P. and J.C. have accessed and verified the underlying data.srhmph@who.int. Data sharing with any individual or organization will be subject to WHO data sharing policy.Request for access to these data can be made to the World Health Organization through All authors declare no competing interests.


Are late preterm birth and size for gestational age associated with cardiometabolic risk (CMR) in childhood?In this cohort study including 1742 children aged 3 to 12 years, those born late preterm and moderately preterm had significantly higher CMR compared with those born full term.This study’s findings suggest that because the CMR score tracks risk from childhood into adulthood, late preterm and moderately preterm birth may be important risk factors for cardiometabolic disorders later in life. The long-term cardiometabolic consequences of late preterm birth (34-36 weeks’ gestation) are not well understood.To assess whether late preterm birth and size for gestational age are associated with cardiometabolic risk (CMR) in childhood.This retrospective cohort study included 1742 children born in Ontario, Canada, between April 1, 2006, and September 30, 2014, and followed up until September 30, 2019. Data from children enrolled in The Applied Research Group for Kids (TARGet Kids!) primary care practice–based research network were linked to administrative health care data at ICES . Participants were excluded if they had conditions affecting growth , any acute or chronic conditions (other than asthma and high-functioning autism), severe developmental delay, or families who were unable to communicate in English.Late preterm birth, gestational age as a continuous measure, and size for gestational age.z score of CMR components, including waist circumference, log triglyceride level, glucose level, systolic blood pressure, and high-density lipoprotein cholesterol level). Secondary outcomes were the individual CMR components. Multivariable linear regression analysis was used to separately evaluate the associations of late preterm birth, continuous gestational age, and size for gestational age with CMR at ages 3 to 12 years.The primary outcome was composite CMR score  were included in the final cohort. Overall, 87 children (5.0%) were born moderately preterm (<34 weeks’ gestation), 145 (8.3%) were born late preterm (34-36 weeks’ gestation), 455 (26.1%) were born early term (37-38 weeks’ gestation), and 1055 (60.6%) were born full term (≥39 weeks’ gestation). Compared with children born full term, those born moderately preterm  and late preterm  had higher CMR scores. Each additional gestational week was associated with a 0.06 U  decrease in CMR.In this study, children born late preterm and moderately preterm had higher CMR. These results suggest that screening and early-life interventions for these children may prevent cardiometabolic outcomes. This cohort study assesses whether late preterm birth and size for gestational age are associated with cardiometabolic risk in childhood among participants aged 3 to 12 years. Late preterm birth (34-36 weeks’ gestation) now accounts for 75% of all preterm births, and this proportion has been steadily increasing since 1990.2 Increases in obstetrical interventions, such as induced and cesarean delivery, account for part of the increase in late preterm births, but much of the reason for this increase remains unexplained.5 In addition, risk factors for spontaneous preterm delivery, including older maternal age, multiple-gestation births, maternal obesity, and maternal diabetes, have also been increasing.7 There is growing awareness that late preterm newborns are not as physiologically or metabolically mature as term newborns and that late preterm birth may confer negative consequences across a child’s life course.8 Size for gestational age, a measure that incorporates infant size at birth, has also been acknowledged as having an impact for health later in life through altered childhood growth and overweight status.10It is well established that early-life exposures during the intrauterine and early postnatal period are associated with risk of adult-onset chronic diseases.11 examining the association between late preterm birth and cardiometabolic health outcomes suggested that children younger than 18 years who were born late preterm vs term were at increased risk of diabetes  and hypertension . Studies13 have also found that adults born late preterm compared with term had high levels of cardiometabolic risk (CMR) factors, including higher body fat percentage, higher blood pressure, metabolic syndrome, and stroke. Small for gestational age (SGA) status has been associated with cardiometabolic outcomes in adulthood,14 and children born SGA and large for gestational age (LGA) are at increased risk of overweight and obesity in childhood and adulthood.10 It is well established that CMR factors do not occur in isolation,15 and the use of a continuous CMR score may better capture potential disease risk. However, the association between CMR score and late preterm birth has not yet been investigated. Among children, CMR is commonly defined as a continuous risk score based on the clustering of age- and sex-standardized central adiposity, high serum lipid levels, glucose levels, and blood pressure.16 The objectives of this cohort study were to evaluate whether late preterm birth, gestational age as a continuous measure, and size for gestational age were associated with CMR among children aged 3 to 12 years who were participating in a cohort study in Toronto, Canada.A systematic reviewSTROBE) reporting guideline for cohort studies.We conducted a retrospective cohort study of children participating in The Applied Research Group for Kids (TARGet Kids!) primary care practice–based research network in Toronto, Canada. Data from children in TARGet Kids! who had CMR outcomes measured at ages 3 to 12 years between April 1, 2006, and September 30, 2019, were individually linked to health care administrative databases at ICES  in Ontario, Canada. Ontario has a universal health care system through which care is provided to almost all residents. ICES is an independent nonprofit research institute whose legal status under section 45 of the Ontario Personal Health Information Protection Act allows it to collect and analyze health care and demographic data without requiring individual participant consent for health system evaluation and improvement. Ethics approval was granted by the Hospital for Sick Children, Unity Health Toronto, and the Hamilton Integrated Research Ethics Boards. All parents of TARGet Kids! participants provided written consent for their child’s data to be linked to health administrative data. This study followed the Strengthening the Reporting of Observational Studies in Epidemiology , any acute or chronic conditions (other than asthma and high-functioning autism), severe developmental delay, or families who were unable to communicate in English.17 Birth at less than 32 weeks’ gestational age (considered very preterm) was a TARGet Kids! exclusion criterion at enrollment. However, some parents or caregivers did not know their child’s gestational age, and these children were included in the analysis; thus, a small number of children with gestational age less than 32 weeks were included. The flowchart of study participants is provided in the TARGet Kids! is an open longitudinal cohort study established in 2008.19 a maternal-newborn data set that identifies hospital discharge records from the Canadian Institute for Health Information Discharge Abstract Database.20 The ICES databases include perinatal data for children born between April 1, 2006, and September 30, 2014, with linkages to maternal data. Those data were also linked to the Better Outcomes Registry and Network (BORN) using unique encoded identifiers; BORN is a provincewide registry of all births in Ontario, Canada, and is part of the Children’s Hospital of Eastern Ontario. Data in this registry are routinely collected from medical records, clinical forms, and patient interviews.21 A validation study found greater than 90% agreement between BORN and patient medical records for most of the audited variables.21 Data from BORN were used to identify home births and measure maternal body mass index  and pregnancy complications among all participants.Data from TARGet Kids! were linked (with 98% success) with the ICES Mother-Baby database,22 Several disease registries, including the Ontario Diabetes Database (which includes patients with physician-diagnosed diabetes) and the Ontario Hypertension Database (which includes patients with physician-diagnosed hypertension),22 which were derived from the health administrative data sets described in the previous paragraphs, were used to measure chronic conditions among mothers.The cohort was further linked with the Canadian Institute for Health Information Discharge Abstract Database to measure hospital admissions and with the Ontario Health Insurance Plan database to measure outpatient visits.23 Gestational age was analyzed as a categorical variable with 4 groups defined a priori consistent with clinical definitions: moderately preterm (<34 weeks’ gestation), late preterm (34-36 weeks’ gestation), early term (37-38 weeks’ gestation), and full term .25 Secondary analyses evaluated gestational age as a continuous variable. Size for gestational age was investigated using established cutoffs because of the documented U-shaped association between size for gestational age and cardiometabolic disease risk.10 Size for gestational age was derived by applying the 1999 reference birth weight curves for the Canadian population to the present study population.26 Size for gestational age was analyzed categorically using the following 3 groups: SGA (<10th percentile), appropriate for gestational age , and LGA (>90th percentile).The primary exposure was gestational age at birth, ascertained from the Mother-Baby database and supplemented with data from BORN to capture home births. Gestational age at birth was calculated based on the best clinical estimate of gestation, including estimates from both ultrasonography and last menstrual period.z score of these components by summing waist circumference, log triglycerides, glucose, and SBP and subtracting HDL cholesterol, then dividing this value by the square root of 5.27 The latest (oldest age) measure of CMR was recorded for children with multiple measurements. The distribution of CMR scores is provided in eFigure 1 in the The primary outcome was CMR score at ages 3 to 12 years identified from TARGet Kids! data. The individual CMR components  were analyzed as secondary outcomes. The overall CMR score was calculated as the total age- and sex-standardized 28The CMR score was only calculated for children with data available for all components. This score was analyzed continuously, and the associations between the exposures and each of the CMR components were also assessed individually. Nonfasting blood samples, child’s waist circumference, and SBP were measured by trained research staff members using standardized instruments.11 and conceptualized using a directed acyclic graph  and fully adjusted (model 2) to assess how the confounding variables changed the observed effect estimates. Model 1 was adjusted for maternal age at delivery, maternal ethnicity, child’s age when the outcome was measured, child’s sex, and annual family income. Model 2 was adjusted for all confounding variables in model 1 plus maternal prepregnancy BMI, maternal prepregnancy cardiometabolic conditions (diabetes and hypertension), gestational CMR , and family history of cardiometabolic conditions. When analyzing individual CMR components, we also controlled for child’s fasting time  at blood sample collection when conducting the multivariable linear regression analysis for the individual outcome of glucose level and controlled for child’s height for the individual outcome of SBP.28Multivariable linear regression analysis was used to estimate β coefficients and 95% CIs for the associations of gestational age at birth with overall CMR and CMR components, which were analyzed individually. The a priori level of significance was 2-tailed Missing data for the covariates were considered missing at random and imputed using multiple imputation with a total of 10 imputations using the PROC MI procedure in SAS software, version 9.4 (SAS Institute Inc). Continuous variables were imputed via predictive mean matching using 100 iterations. Categorical variables were imputed using the Markov chain Monte Carlo method, which assumes that all variables in the imputation model have a joint multivariable normal distribution. There was less than 15% missingness for each variable before imputation . Consistz score for the total cohort was −0.03 (1.20). Among mothers, the mean (SD) age was 33.6 (4.5) years. A total of 238 mothers (13.7%) were of African, Arab, Latin American, or multiple ethnicity; 315 (18.1%) were of East Asian, South Asian, or Southeast Asian ethnicity; 1028 (59.0%) were of European ethnicity; and 161 (9.2%) were of unknown ethnicity (missing data).Among 2440 eligible children, 1742  were included in the final cohort ; Figure Children born late preterm vs full term were more likely to be male (86 of 145 children [59.3%] vs 565 of 1055 children [53.6%]) and have lower annual family income (<$50 000: 21 of 145 children [14.5%] vs 66 of 1055 children [6.3%]); their mothers were more likely to be older  and of non-European ethnicity (51 of 131 mothers [38.9%] vs 312 of 959 mothers [32.5%]) and to have gestational diabetes (7-11 mothers [4.8%-7.6%] vs 30 mothers [2.8%]), gestational hypertension (7-11 mothers [4.8%-7.6%] vs 23 mothers [2.2%]), or preeclampsia or eclampsia (10-14 mothers [6.9%-9.7%] vs 1-5 mothers [0.1%-0.5%]). The percentage of missing values across the covariates ranged from 0%  to 14.5% . Additional characteristics of the study population, including CMR outcome score by gestational age categories, are shown in The minimally adjusted (model 1) and fully adjusted (model 2) associations between gestational age and CMR are shown in For the individual CMR components, compared with full-term birth, late preterm birth  and moderately preterm birth  were associated with higher SBP based on the fully adjusted model . CompareThe secondary analysis evaluating the association between size for gestational age and CMR revealed no association between SGA status and overall CMR  in the fully adjusted model. No association was found between LGA status and overall CMR  compared with AGA status . For the13 earlier gestational ages were associated with adverse cardiovascular outcomes. Both late and moderately preterm birth were associated with higher overall CMR scores, higher SBP, higher triglyceride levels, and lower HDL cholesterol levels. Being born LGA was associated with higher waist circumference, whereas being born SGA was associated with lower waist circumference. Some data suggested that LGA status may have been associated with higher overall CMR scores, but these results were not statistically significant.In this retrospective cohort study, late preterm birth was associated with higher CMR scores among children aged 3 to 12 years. Consistent with previous literature,34 that reported associations between preterm birth and CMR. However, those reviews did not find consistent differences in adulthood between those born preterm vs term for all components of the metabolic syndrome, possibly owing to improper adjustment, specifically overadjustment for variables on the causal pathway, such as birth weight. We did not adjust for variables on the causal pathway to ensure that the main exposure estimates were not biased. In a retrospective cohort study of 720 children,13 early preterm birth  and late preterm birth  were associated with metabolic syndrome  in adolescence.Our results were consistent with several reviews36 found that both SGA status and LGA status were associated with higher CMR. In a prospective cohort study involving 90 children in Italy, Chiavaroli et al35 found that children born LGA and SGA had a higher CMR score compared with children born AGA, with increased differences at adolescence. Kuhle et al36 analyzed data from the Canadian Health Measures Survey to find that children born SGA were less likely to have high levels of 3 or more components of metabolic syndrome at ages 6 to 12 years compared with their peers born AGA. A small review by Nordman et al10 that investigated the association between size for gestational age and CMR factors reported no differences in fasting glucose levels among children born SGA and LGA compared with those born AGA but did find increased insulin levels and insulin resistance. Both SGA and LGA status were found to be associated with high blood pressure in children and/or adolescents.10 However, the review suggested that sex and ethnicity could explain some of the variation among those born LGA.10 Overall, there appears to be evidence that the association between size for gestational age and cardiometabolic disease risk may be U-shaped, with children born SGA and LGA having increased risk of some CMR factors, which could be explained by catch-up growth and catch-down growth, respectively, during the first years of life.10Previous studies37 The mechanisms associated with hypertension and cardiovascular disorders are believed to involve changes to the vasculature and heart ,38 kidneys ,39 and sympathetic nervous system.37 Mechanisms associated with abnormal metabolic homeostasis in those born preterm are complex and remain unclear but could involve changes to the vasculature 38 and nutritional deficits both in utero and after delivery.37 Catch-up growth in weight without a parallel catch up in length may then result in an obesogenic phenotype similar to that of infants born SGA.37 Children born LGA may compensate for higher intrauterine growth through catch-down growth in early years; however, lack of catch-down growth seems to increase the risk of cardiometabolic outcomes in LGA-born young adults.10These findings may have multiple underlying mechanisms; adverse conditions, both in utero and after birth, occur at a critical stage of organ system development, which could lead to permanent alterations through epigenetic and genetic mechanisms.These findings may have clinical implications. First, they suggest that late preterm birth and possibly LGA status may be important risk factors for cardiometabolic disorders. Because the CMR score tracks risk from childhood into adulthood, early preventive evaluation and CMR monitoring beginning early in childhood is warranted for preterm-born children. To identify these children, medical records and history, including birth history , need to be routinely collected for all infants; this information is already regularly collected in clinical practice. In addition, timely treatment options could be developed for children with signs of CMR to reduce subsequent cardiometabolic outcomes.37Advances in neonatal and pediatric care in the past few decades mean that a substantial proportion of children born late preterm are now reaching young to mid-adulthood. Thus, there is a need to understand and prevent the negative cardiovascular health consequences of late preterm birth through targeted interventions before this population reaches older adulthood, when cardiovascular disease incidence increases. Children with a history of preterm birth may need early preventive evaluation and long-term monitoring for cardiometabolic outcomes later in life. Future studies could evaluate whether cardiovascular risk screening for children born preterm improves outcomes among children and young adults and whether population health interventions targeted at children born both early and late preterm mitigate adverse outcomes. More data from larger studies in diverse populations are needed to understand whether recommended perinatal interventions improve preterm birth outcomes and modify the association between late preterm birth and CMR.40 allowing for earlier intervention. Extremes of in utero growth, maternal factors, and complications during pregnancy may lead to the decision to induce preterm labor and may also be independently associated with increased CMR later in life. However, a strength of our study was that we were able to control for these potential confounding variables by combining administrative medical records data, cohort study questionnaire data, and physical measures to find that preterm birth was associated with higher CMR in childhood.This study has several strengths. A notable strength was the ability to link clinical data from a large cohort of children to health administrative data from a publicly funded health care system to obtain valid prospectively measured data on gestational age at birth as well as maternal factors and health conditions. Components of the CMR score were measured objectively in childhood by trained staff, and the approach we used to calculate CMR is emerging as a preclinical (or intermediary) cardiometabolic outcome measure among children at risk of developing cardiometabolic disease in adulthood,The study also has limitations. We cannot rule out the possibility of residual confounding. We have adjusted for height in our SBP models, but national reference standards to allow standardization by age, sex, and height may be necessary for blood pressure measurement in children. Although there were generally few missing variables, missingness for maternal BMI was 14.5%. To address the presence of missing data, multiple imputation was used. Another limitation is that our study was nested within the TARGet Kids! primary care practice–based research network and reflects children recruited from selected primary care practices rather than a population-based representative sample. The findings from this study may not be generalizable to other populations, and selection bias is possible. Children in the study sample were from families that had a relatively high income compared with the overall population.In this cohort study, late preterm and moderately preterm birth were associated with higher CMR. Because the CMR score tracks risk from childhood into adulthood, late preterm and moderately preterm birth may be important risk factors for cardiometabolic disorders later in life. Screening and early-life interventions for these children may prevent cardiometabolic outcomes.


This study aimed to provide additional information on the influence of perinatal factors on fecal (f)-calprotectin values in preterm infants. Calprotectin was determined from the first spontaneous stool  obtained from neonates at a mean age of 3.41 ± 2.44 days of life. We analyzed 114 subjects who had a body weight of 1847.67 ± 418.6 g and were born at a gestational age of 32.6 ± 2.43 weeks, without intestinal and other congenital anomalies or any diseases other than those related to premature birth. The values of f-calprotectin are in a positive correlation with female subjects, intrauterine growth restriction, significant ductus arteriosus, enteral feeding intolerance, postnatal prolonged use of broad-spectrum antibiotics, and values of bicarbonates . Measurement of f-calprotectin in the first 7 days after birth can help to early detect the intestinal distress or early staging of necrotizing enterocolitis in premature infants. The granulocytes excrete calprotectin actively into the lumen of the digestive tract ,2,3,4,5.2α) (0.05). The required sample size in the study group is at least 61 subjects.The sample size was calculated “a priori” (G-power software), based on the following baseline parameters: study power (80%), clinically significant value of Pearson’s correlation coefficient (0.35) and first type error  hospitalized at the Neonatal Intensive Care Unit (NICU) of the Center for Neonatology, the University Clinical Center Kragujevac, Serbia. The inclusion criteria were as follows: gestational age of <37 weeks, postnatal age of <7 days, and the absence of any disease other than those related to premature birth . Excluding criteria were death in the first 24 h of life, gestational age of <24 weeks, postnatal age of ≥7 days at the time of stool collection, gastrointestinal and other congenital anomalies, chromosomal aberrations, and congenital metabolic diseases.® strips, range 0–1,000 µg/g) performing the enzyme-linked immunosorbent assay (ELISA) [After admission to the NICU, all the preterm infants were fed a reconstituted formula for preterm infants, as they were transported from 12 distant maternity hospitals. Calprotectin was determined from the first spontaneous thick, greenish-black and sticky stools after birth. Stools were collected once a day until 10 am from the infants at a mean age of 3.41 ± 2.44 days of life. Stool samples were not stored in the refrigerator. The duration of meconium passage was significantly prolonged in infants with very preterm (28–32 weeks), and especially, extremely preterm birth (less than 28 weeks). The values of f-calprotectin were analyzed immediately after taking the stool samples to the laboratory to be tested on the Alegria device, (24 Alegria (ELISA) ,15. BlooDuring the study period, all preterm infants underwent echocardiography on the third day of life. Significant ductus arteriosus was defined ,17,18,192.2U test. Spearman’s rank correlation coefficient was used for correlating f-calprotectin and continuous variables. A value of p < 0.05 was considered statistically significant.Statistical analyses were performed using SPSS version 18.0 . The continuous variables were presented as the mean value and standard deviation (SD) or median and interquartile range (IQR). The categorical data were presented as the number and proportion (%). Comparisons of f-calprotectin according to clinical factors were performed by Mann–Whitney 3The demographic characteristics of the patients included in the study are shown in 2), and carbon dioxide (pCO2) analyzed in a sample of capillary arterial blood were not. Significantly higher fecal calprotectin were detected in female newborn (p = 0.003) significant ductus arteriosus (p = 0.025), postnatal prolonged use of broad-spectrum antibiotics (p = 0.040), and vomiting (p = 0.038)   and 3.p = 0.038).Significantly higher fecal calprotectin levels  were detected in subject with feeding intolerance (performed due to vomiting) compared with the group without enteral feeding intolerance , serum procalcitonin, urea, creatinine, and positive blood culture did not significantly affect its values .4The gastrointestinal tract (GIT) of the newborns is considered sterile at birth. After birth, it becomes rapidly colonized by microorganisms from the birth canal or from the environment. According to the recent hypothesis, the GIT of the neonate is first inhabited by the microorganisms from the mother’s GIT and uterus, followed by the colonization of the bacteria present in mother’s milk. The mechanism is still unclear. Bacterial colonization of the GIT of the newborn can be considered a kind of the fetus’ immune response to adapt to life outside the uterus . Today, During the study period, significantly higher fecal calprotectin levels were recorded in female newborns with intrauterine growth restriction (IUGR) than in the group of males without IUGR. Certain studies ,22,23 reIn the era of personalized medicine, non-invasive biomarkers can play a key role in reducing neonatal mortality, first and foremost, by enabling more accurate assessment of the risk for the development of neonatal diseases, more appropriate therapeutic treatment, and earlier prediction of the clinical outcome. Previous research suggested that the determination of the calprotectin in the stool of preterm infants could be crucially important since it has already been validated in adults and children as a sensitive marker of inflammatory bowel disease (Crohn’s disease and ulcerative colitis . F-calprDue to large inter- and intra-individual variations, the precise determination of the limit values for f-calprotectin remains an unachievable target, and the proposed limit values, depending on the author, range from 200 to 2,000 μg/g ,22. DuriVomiting, compensatory elevated values of bicarbonates, hemodynamically significant ductus arteriosus, and prolonged postnatal treatment of broad-spectrum antibiotics were associated with higher values of fecal calprotectin, as a sign of intestinal distress or GIT inflammation ,28. It iEnteral feeding intolerance is very common in preterm infants ,28 and cNEC almost exclusively occurs in preterm infants followed by the development of intestinal inflammation with a significant participation of neutrophils, because of the immature enterocyte response to bacterial stimulation and the onset of oral feeding. While some authors find that f-calprotectin levels increase significantly in VLBW premature infants with NEC , other a5New findings on calprotectin as a potentially useful non-invasive parameter of the intestinal distress in preterm infants can be of great importance in the era of a personalized medical approach to the newborns. The diagnostic accuracy of f-calprotectin measurement in preterm infants can be confirmed and increased by including this parameter in the panel with other biomarkers, as well as with the better understanding of the factors that affect its excretion. Our results show that an increase in f-calprotectin, in a population of preterm infants, correlates with IUGR, the enteral feeding intolerance, significant patent ductus arteriosus, and prolonged antibiotic therapy. Measurement of f-calprotectin in the first 7 days after birth can help to detect early the intestinal distress or early staging of NEC in premature infants.


P = 0.037). Kaplan–Meier curve analysis revealed that freedom from MACEs exhibited a greater decrease in the high apo-C3 group than in the low apo-C3 group, and in the high %NC group than in the low %NC group. Multivariate Cox hazards analysis showed that the %NC and high apo-C3 were independent predictors of 4 year MACEs. Apo-C3 may be a useful marker of future MACEs in patients with SCD after PCI and contribute to %NC growth.We aimed to clarify the relationship between apolipoprotein C3 (apo-C3) and the vascular composition of lesion plaque in stable coronary disease (SCD) before percutaneous coronary intervention (PCI), and to investigate major adverse cardiovascular events (MACEs) within 4 years. Data of 98 consecutive patients with SCD who underwent PCI between November 1, 2012, and March 10, 2015, were analyzed. Laboratory and virtual histology-intravascular ultrasound (VH-IVUS) examinations of culprit lesions were conducted before PCI. Patients were divided according to median apo-C3 into low apo-C3 (≤ 8.5 mg/dL) and high apo-C3 (> 8.5 mg/dL) groups. VH-IVUS data indicated that the percentage of necrotic core volume (%NC) was significantly higher in the high apo-C3 group than in the low apo-C3 group. Moreover, the %NC significantly correlated with the apo-C3 level (R = 0.2109,  Previously, we showed the correlation of the %NC with apolipoprotein B and low-density lipoprotein (LDL) cholesterol5. These effects are presumed to be partly due to the lowering of LDL cholesterol. However, previous reports7 have shown that the cholesterol-lowering effect of strong statins on major adverse cardiovascular events (MACEs) was only 24% when compared with that of placebo. Thus, LDL cholesterol may not be the only factor that affects plaque volume and vessel composition.In a meta-analysis13. Apo-C3 concentrations in chylomicron-free serum were reported to be associated with event-free survival in patients with coronary artery disease13. Nonetheless, thus far, the relationship between apo-C3 and vessel composition has not been clarified. In particular, NC is a key feature of vulnerable plaque14, and MACEs associated with NC have not been investigated.Apolipoprotein C3 (apo-C3) is associated with triglyceride-rich lipoprotein metabolism and has emerged as an independent marker of the risk for cardiovascular disease (CVD)Therefore, in this study, we conducted VH-IVUS before PCI to examine the vessel structure and to evaluate the relationship between apo-C3 and vessel composition. We also analyzed the Kaplan–Meier curves for MACEs associated with apo-C3 or the %NC in the plaque in patients with stable coronary disease (SCD) after PCI.During the follow-up period of 48 months, 26 MACEs occurred , 5 restenoses of the PCI site (including 4 lesions requiring PCI) and two arteriosclerosis obliterans requiring endovascular therapy , triglycerides (TG), non-high-density lipoprotein cholesterol (HDL-C), apo-A1, apo-B, and apo-C3 were higher in the high apo-C3 group than in the low apo-C3 group Table . The LDLComparisons of grayscale IVUS, VH-IVUS data, and MACEs between the low and high apo-C3 groups are presented in Table Grayscale IVUS data, including %plaque burden, were all comparable between the low and high apo-C3 groups. Regarding VH-IVUS data, the percentage of the fibrofatty (FF) volume in the plaque volume (%FF) was significantly lower in the high apo-C3 group than in the low apo-C3 group. Meanwhile, the %NC volume was significantly higher in the high apo-C3 group than in the low apo-C3 group. Total MACE and the incidence of new lesions were significantly higher in the high apo-C3 group than in the low apo-C3 group.P = 0.037)  and percentage of the DC volume in the plaque volume (%DC) . The %FF significantly inversely correlated with the %NC  and %DC . The %NC significantly positively correlated with the %DC . A matrix showing the correlations between lipid levels, apolipoprotein levels, and plaque components is shown in Supplementary Table Furthermore, we investigated the correlations between apo-C3 levels and four vessel compositions evaluated by VH-IVUS. Only the %NC among the four compositions significantly correlated with the apo-C3 level  had significantly lower freedom from MACEs than those with apo-C3 levels ≤ 8.5 mg/dL  (P = 0.0026). The mean survival time with high apo-C3 was 33.3 ± 2.7  months, while that with low apo-C3 was 44.4 ± 1.4 (95% CI 41.7–47.1) months. Interestingly, patients with high %NC  of the lesion plaque showed significantly lower freedom from MACEs than those with low %NC  of the lesion plaque , %DC , and high apo-C3  were significant risk factors for 4 year MACEs. Moreover, the multivariate analysis showed that the %NC  and high apo-C3  were strong predictors of 4 year MACEs (Table P = 0.0197) and high apo-C3  could predict 4 year MACEs. Similarly, in a model adjusted for apo-A1 and estimated glomerular filtration rate (eGFR) (model 2), %NC  and high apo-C3  were independent predictors of 4 year MACEs  were independent predictors of 4 year MACEs after adjustment for LDL-C and apo-B (model 1). In model 2, adjusted for apo-A1 and eGFR, high apo-C3  was an independent predictor of 4-year MACEs  and %NC  but altered the proportion of the composition of blood vessels. Apo-C3 decreased the %FF and increased the %NC. The NC may be formed with a matrix-devoid gruel of lipids and cell debris absorbed by macrophages and dendritic cells in the intima in a process that remains incompletely understood in vivoSince the %NC cannot be assessed without VH-IVUS, which is an invasive method, it may be possible to predict future myocardial infarction if apo-C3, which can be assessed by blood sampling, is checked regularly as a surrogate marker in the outpatient clinic.This study has some limitations. First, this was a single-center study with a small sample; therefore, large-scale multicenter studies are needed to confirm our findings. However, with the advent of high-resolution IVUS, only VH-IVUS is used for clinical research purposes now, making it more difficult to conduct a study, since it is used less frequently. The possibility that unknown confounding factors might have influenced the results cannot be completely denied, even though multivariate analyses adjusted for known prognostic factors. However, there is no drug bias, as the oral medication was virtually unchanged during the study period. Furthermore, the correlation between apo-C3 and %NC volume of pre-PCI plaques is clear. The %NC volume of plaques with new lesions in the high apo-C3 group was as high as 20.6% at 9–12 months after the first PCI and was much higher than those of pre-PCI plaques. These results support the fact that high apo-C3 increases %NC and creates new lesions.Second, because of the small sample size, we could not examine additional confounding factors for apo-C3. Finally, most of the new lesions were discovered on repeated CAG at 9–12 months after PCI. Nowadays, it is difficult to obtain informed consent to perform CAG 9–12 months after PCI. However, in routine clinical practice, new lesions cannot be traced.In conclusion, apo-C3 facilitates NC progression in the coronary arteries of patients with SCD and increases the risk of future cardiovascular events. Apo-C3 may be a potent surrogate marker for predicting future cardiovascular events in patients with SCD after PCI. Apo-C3 levels should be monitored periodically in outpatient medical examinations. If apo-C3 level is high, closer examination using coronary computed tomography, stress test, or CAG should be performed to monitor CVD progression.This study was approved by the Ethics Committee of Fukushima Red Cross Hospital  and performed in accordance with the Ethical Guidelines for Clinical Research of the Japanese Ministry of Health, Labour and Welfare. Informed consent was obtained from all individual participants included in the study.25. After obtaining written informed consent, PCI and pre-IVUS were performed. All 98 patients were enrolled in this study, all of whom underwent coronary artery stenting in the target vessels. The degree of coronary artery stenosis was determined by consultation between two or more experienced interventional cardiologists . IVUS measurements were assessed by two investigators (TO and TS) who were blinded to the treatment allocation.Our sample included a total of 98 consecutive patients with SCD who visited our hospital between November 1, 2012, and March 10, 2015, and underwent PCI with VH-IVUS. PCI was indicated in accordance with the Japanese consensus criteriaIVUS examinations of the culprit lesions were conducted immediately before PCI. A phased-array, 20 MHz, 3.2-F IVUS catheter  was placed in the distal coronary artery and pulled back to the aorto-ostial junction using a motorized catheter pull-back system set at 0.5 mm/s . The grayscale IVUS and captured radiofrequency data were recorded by a DVD recorder.26. In the gray-scale conventional IVUS analysis, images were assessed to determine the lesion length, lumen volume, plaque volume, and vessel volume. In the VH-IVUS analysis, the FI, FF, NC, and DC regions were color coded. Estimates of each factor’s contribution to the volume of the entire culprit lesion plaque volume were reported as a percentage of the factor’s volume in the plaque volume of the culprit lesion 27. The same investigators (TO and TS) subsequently reanalyzed the images with respect to the intra- and inter-observer reproducibility of the measurements.Offline grayscale and VH-IVUS analyses were conducted using echo Plaque 4.0 software . Corresponding proximal and distal reference IVUS images were identified for each culprit lesion and analyzed to determine the plaque volume within the involved arterial segment. Grayscale IVUS analysis was conducted using the American College of Cardiology Clinical Expert Consensus Document on Standards for Acquisition, Measurement, and Reporting of Intravascular Ultrasound StudiesLaboratory evaluations of plasma lipid and apo levels were conducted within 2 days before PCI. Commercial reagent kits  were used to analyze total plasma cholesterol, LDL-C, HDL-C, and TG levels. Lipid concentrations were measured using an automated biochemical analyzer . Apo-C3 was measured in chylomicron-free samples, and apo-B and a were measured using immunonephelometry. High-sensitivity C-reactive protein levels were measured using nephelometry. Apo levels were measured at a central clinical biochemistry laboratory .MACEs within 4 years (48 months) were defined as a composite of cardiac death, acute myocardial infarction, unstable angina, coronary revascularization for restenosis of a target vessel and new lesion, and endovascular therapy for arteriosclerosis obliterans. These events were ascertained by reviewing patients’ medical records and confirmed by direct dialog with patients, their families, and their physicians after discharge. If no events were experienced in about 8 months after PCI, the patients underwent repeated CAG at 9–12 months from first PCI. For > 90% stenosis in a target vessel  or a new lesion, patients underwent revascularization using PCI; these events constituted MACEs.t-test. All statistical analyses were two-sided. The significance level was set at P < 0.05. The comparisons of grayscale IVUS, VH-IVUS data, and MACEs between the low and high apo-C3 groups were performed in the same manner. In some patients, the comparisons of VH-IVUS data between the new lesion plaque at approximately 9–12 months after the first PCI and the plaque of his own which had angiographically non-significant stenosis at pre-first PCI were performed by paired t-test.Data are presented as mean ± standard errors. Categorical and continuous variables of clinical characteristics were compared between the low and high apo-C3 groups, which were divided by the median apo-C3 value. Categorical data were analyzed using the chi-square test, and continuous data using Student’s To allow for a more detailed analysis, the MACEs within 4 years-follow up were divided into whether they were new lesions (new plaque) or restenoses (same plaque); and characterized as requiring endovascular therapy or leading to AMI, UAP, or cardiac death.  were evaluated . Specifically, %NC and high apo-C3 were estimated by multivariable Cox proportional hazards regression models adjusted for major confounding variables such as LDL-C and apo-B (model 1), and apo-A1 and eGFR (model 2).A Cox proportional hazards regression model was used to estimate the long-term MACE HR and 95% CIs based on the univariate analysis of clinical, laboratory, and IVUS data and multivariate analysis of high apo-C3 levels adjusted for potential confounders , and apo-A1 and eGFR (model 2).All statistical analyses were performed using Ekuseru-Toukei for Windows version 1.02 .Supplementary Information.


The diagnosis of ovarian mass during pregnancy may be missed if early ultrasonographic assessments of pregnancy are not done, resulting in late detection and management. Hence, this case report is aimed at underscoring the possibility of misdiagnosis of a giant ovarian mass for massive ascites in pregnancy. We present the case of a 35-year-old unbooked multigravida woman who was referred to our centre on account of polyhydramnios in labour, misdiagnosed as a massive ascites on admission. The incidental intraoperative diagnosis of a right giant ovarian cyst resulted in the right salpingo-oophorectomy of a huge cyst with the histological diagnosis of a benign ovarian serous cystadenoma. Huge cystic ovarian tumours coexisting with a normal pregnancy may simulate massive ascites resulting in a misleading diagnosis as reported in this case study. We report this case to awaken the suspicion index of a giant ovarian cyst in any patient presenting with massive ascites in pregnancy. The occurrence of coexisting ovarian cysts in pregnancy is not uncommon in pregnancies where there are acceptable diagnostic imaging techniques . The broOvarian cysts are diagnosed in 0.05-5% of pregnancies , which aOvarian cysts coexisting with pregnancy are usually asymptomatic and incidentally detected during ultrasound assessment of early trimester pregnancy, which may resolve spontaneously . However Few reported cases of giant ovarian cysts were misdiagnosed as ascites -10. ThisWe present the case of a 35-year-old female  at an estimated gestational age of 38 weeks and six days. She was a trader with a primary level of education, referred from Civil Service Hospital, in Ilorin, Nigeria .Upon referral, she was admitted to the labour room of General Hospital, Ilorin, Nigeria with complaints of intermittent lower abdominal pain and liquor drainage per vaginum three hours before presentation. The indication for referral was ultrasound confirmation of polyhydramnios in labour with attending potential foetal risk and need for neonatal care. She was unbooked and presented at the referring centre with an ultrasound result showing severe polyhydramnios at term (amniotic fluid index of 46 centimetres) with sudden onset of liquor drainage resulting in her transfer. Her past obstetrics and medical histories were normal.On general examination, her vitals were stable. The abdomen was grossly distended with difficulty in delineating the foetal parts and palpating for uterine contraction. There was relative palpable fundal height at 34 centimetres with a positive fluid thrill in the upper quadrants of the abdomen above the fundus of the uterus and flanks fullness raising suspicion of severe ascites. The foetal heart sound heard with foetal Sonicaid  was 148 beats per minute. From pelvic examination, she was in the active phase of first-stage labour on admission with cervical dilatation of 6cm, 80% effacement, vertex presentation, station 0, and absent foetal membranes.Laboratory tests including complete blood count, electrolytes, urea and creatinine, random blood sugar, and viral marks (hepatitis and HIV) were within normal limits. A repeat obstetric ultrasound scan on admission revealed a viable foetus with massive ascites.She had augmentation of labour with oxytocin infusion but had poor progress of labour despite adequate contraction and, hence, underwent emergency caesarean section under region anaesthesia. An alive female baby of 2.7 kg was delivered with good APGAR scores at the first and fifth minutes. There were no noticeable ascites but a giant right-sided ovarian cyst extending to the epigastric region after exteriorizing the uterus. The uterine incision was closed in two layers with absorbable Vicryl 2. An attempt at exteriorizing the cyst through the Pfannenstiel incision was impossible.The cyst tension was decompressed after suctioning about three litres of cyst contents  using a venous cannula needle perforation (cystostomy) for ease of exteriorizing the cyst through the Pfannenstiel incision site. The cyst wall was elevated with two clamps to prevent leakage of the cyst fluid. The needle was then removed and an incision of 0.5cm in diameter was made at the same place. An aspirator tip was placed through that incision to reduce the cyst size and the 0.5cm wide cut that was made to the cyst wall was ligated. When the authors were aspirating the cyst fluid, controlled reperfusion was performed by the anaesthesiology team to prevent persistent hypotension during the operation. The decompressed right giant ovarian cyst exteriorized, a right salpingo-oophorectomy was done, and the sample was sent for routine histological evaluation.The exploration of the lower abdomen showed that the mass was originating from the right ovary. The left ovary and other pelvic organs were normal in appearance. The total weight of the aspirated cyst content and removed cystic mass was 5.7 kg. The abdomen was closed in layers. Figure The postoperative period was uneventful, and the patient was discharged from the hospital on the third day after surgery. No complications were observed in the six weeks after surgery. The histopathological study of the specimen reported a serous cystadenoma of the ovary.The occurrence of giant ovarian cysts in pregnancy simulating ascites has been seldomly reported in the literature ,11. DespOvarian cyst coexisting with pregnancy has various manifestations, often non-specific pressure symptoms, such as back or abdominal pain, constipation, abdominal distension, and urinary complaints . Since tSome patients in our setups seek antenatal care late due to illiteracy, favourable previous pregnancy experiences, and financial constraints but this is not without risks of morbidity and mortality when complications develop. The non-specific clinical features in ovarian cysts mimicking some features of pregnancy lead to the challenges of late diagnosis and delayed intervention if no diagnostic imaging is done .Many giant ovarian cysts can present with signs and symptoms of ascites due to their large nature and they are commonly mistaken for it, with the flanks fullness, and demonstrable shifting dullness on abdominal examination . This waIntraoperative diagnosis of giant ovarian cysts may be inevitable where there are preoperative misdiagnoses, and the inappropriate choice of abdominal incision may create surgical challenges as in this patient.The presence of very large cystic masses in pregnancy may masquerade as ascites resulting in a misleading diagnosis and altering the mode of treatment. The ultrasonographic examination of giant ovarian cysts in pregnancy is not without misdiagnosis if not thoroughly evaluated. Hence, giant ovarian cysts should always be considered in the differential diagnosis of large ascites in pregnancy for appropriate intervention. The role of antenatal care in this instance cannot be overemphasized.


To assess the risk of preterm birth, small for gestational age at birth, and stillbirth after covid-19 vaccination during pregnancy.Population based retrospective cohort study.Ontario, Canada, 1 May to 31 December 2021.All liveborn and stillborn infants from pregnancies conceived at least 42 weeks before the end of the study period and with gestational age ≥20 weeks or birth weight ≥500 g.Using Cox regression, hazard ratios and 95% confidence intervals were estimated for preterm birth before 37 weeks , very preterm birth (<32 weeks), small for gestational age at birth (<10th centile), and stillbirth. Vaccination against covid-19 was treated as a time varying exposure in the outcome specific risk window, and propensity score weighting was used to adjust hazard ratios for potential confounding.v 6.9% among unvaccinated; adjusted hazard ratio 1.02, 95% confidence interval 0.96 to 1.08), spontaneous preterm birth , or very preterm birth . No increase was found in risk of small for gestational age at birth  or stillbirth . Findings were similar by trimester of vaccination, mRNA vaccine product, and number of doses received during pregnancy.Among 85 162 births, 43 099 (50.6%) occurred in individuals who received one dose or more of a covid-19 vaccine during pregnancy—42 979 (99.7%) received an mRNA vaccine. Vaccination during pregnancy was not associated with any increased risk of overall preterm birth (6.5% among vaccinated The findings suggest that vaccination against covid-19 during pregnancy is not associated with a higher risk of preterm birth, small for gestational age at birth, or stillbirth. Infection with SARS-CoV-2 during pregnancy has been associated with higher risks of admission to hospital, admission to an intensive care unit, and death for pregnant individuals.Safety concerns about covid-19 vaccination during pregnancy remains a potential obstacle to improving coverage. As of July 2022, results published from epidemiological studies are reassuring—two case-control studies of covid-19 vaccination in early pregnancy found no association with spontaneous abortion.We followed guidance for conducting studies of covid-19 vaccination during pregnancyWe conducted a population based retrospective cohort study in Ontario, Canada, which provides publicly funded healthcare to all residents, including services relating to prenatal and obstetrical care. The provincial birth registry (Better Outcomes Registry and Network (BORN Ontario))The birth registry receives records for all liveborn infants and stillborn infants ≥20 weeks’ gestation or with birth weight ≥500 g from hospitals, birth centres, and midwifery practice groups (home births) across Ontario.Ontario’s covid-19 vaccination programme began on 14 December 2020,We defined preterm birth and very preterm birth as a live birth before 37 and 32 completed weeks of gestation, respectively. Preterm birth subtype was considered spontaneous if it occurred after spontaneous onset of labour or preterm premature rupture of membranes.v <30), self-reported smoking status (yes or no) or substance use during pregnancy (yes or no), public health unit region (seven regions), pre-existing maternal health conditions , parity (number of previous live births and stillbirths), multi-fetal pregnancy (yes or no), rural or urban residence, neighbourhood income fifths, neighbourhood marginalisation fifths , calendar week of conception , and first visit for prenatal care in the first trimester (yes or no). Supplementary table 2 provides additional details on outcomes and covariates.We adjusted for a range of covariates potentially associated with study outcomes or covid-19 vaccination, or both, using propensity score methods: maternal age at delivery (years), prepregnancy body mass index ≥30 ; ongoing pregnancies on 1 May (start of the study) started contributing time from that point (see supplementary fig 1), and follow-up continued until either the event or censoring at the end of the outcome specific risk window.We used extended Cox proportional hazards regression models with gestational age in days as the time scale, and robust sandwich variance estimation to account for statistical dependence across repeated observations as a result of changes in vaccination status, which was treated as time varying. We estimated hazard ratios with 95% confidence intervals and used inverse probability of treatment weights to generate adjusted hazard ratios.In primary analyses, we evaluated vaccination status as receipt of one dose or more of a covid-19 vaccine in the outcome specific risk window. We performed subgroup analyses for preterm birth and small for gestational age at birth to evaluate trimester specific associations according to the number of doses (treating each dose as time varying), mRNA vaccine product for dose 1, and vaccine product combinations for those who received two mRNA doses; the number of events for the other outcomes was insufficient to conduct these subgroup analyses. In sensitivity analyses, we added covid-19 during pregnancy as a time varying covariate to the model and, separately, excluded individuals with a history of covid-19 during pregnancy. To assess for potential residual confounding, we stratified the original models by neighbourhood income (two highest and three lowest fifths) and limited the unvaccinated group to those who received their first vaccine dose after pregnancy, because in an earlier study of this population their baseline characteristics were shown to be more similar to individuals vaccinated during pregnancy than to those never vaccinated at any time.Owing to covid-19 related resource and time constraints, it was not feasible to directly involve patients in the design, conduct, reporting, or writing of our study. Although no patients were directly involved, the study team included several obstetrical care providers who were involved from the outset of planning the study and brought forward their experiences from patient interactions related to covid-19 vaccination during pregnancy. These were taken into consideration in planning this research and its dissemination to ensure relevancy for a broad group of knowledge users, including pregnant people.v 4.9%; After exclusions, 85 162 live births and stillbirths occurred during the study period ; of thesAmong 43 099 individuals who were vaccinated during pregnancy, 13 416 (31.1%) received one vaccine dose, 29 650 (68.8%) received two doses, and 33 (0.1%) received three doses . OverallOverall, 5719 (6.7%) preterm births occurred; 3450 (4.1%) were spontaneous preterm births . The cumThe cumulative incidence of small for gestational age at birth was 9.1% (n=3743) among births to individuals who received one dose or more of a covid-19 vaccine during pregnancy, and 9.2% (n=3722) among births to unvaccinated individuals. No association was observed between vaccination during pregnancy and small for gestational age at birth overall . Estimates for most outcomes were similar across strata of neighbourhood income; although stratum specific estimates for individuals of higher income (fourth and fifth fifths) and lower income (first to third fifths) were different in magnitude for very preterm birth and for stillbirth, the direction of the stratum specific point estimates was consistent with that of the main analyses, and confidence intervals overlapped. When the comparison group was restricted to unvaccinated individuals who initiated their covid-19 vaccine series after pregnancy, the results for all outcomes remained unchanged. Analyses restricted to singleton births were similar for preterm birth outcomes; however, the estimate for stillbirth moved closer to the null value .In this large population of more than 85 000 live births and stillbirths up to 31 December 2021, we found no evidence that vaccination during pregnancy with an mRNA covid-19 vaccine was associated with a higher risk of preterm birth, spontaneous preterm birth, very preterm birth, small for gestational age at birth, or stillbirth. These results—based on more than 43 000 fetuses exposed to at least one dose of mRNA covid-19 vaccine—did not differ by trimester of vaccination, number of doses received during pregnancy, or mRNA vaccine product.Our findings for preterm birth and small for gestational age at birth are consistent with a recent Scandinavian birth registry based study of 157 521 births (103 409 from Sweden and 54 112 from Norway) to 12 January 2022, in which 18% of the births (n=28 506) were to individuals who were vaccinated during pregnancy.We observed a reduction in stillbirth risk among vaccinated individuals , even after adjustment for a range of potential confounders through propensity score weighting. A reduced risk of stillbirth was also observed in the Scandinavian study, although the estimate was smaller in magnitude .Strengths of this study include the large birth population and number of individuals vaccinated during pregnancy. We used deterministically linked population based data sources within a publicly funded healthcare system; thus, identification of births and vaccinations was highly complete and minimised potential selection or exposure misclassification bias. As gestational age at birth is mostly based on early ultrasound assessment in Ontario, the accuracy of classifying gestational timing of vaccination was probably high. Although we adjusted for many potential confounders using a propensity score based approach, we cannot dismiss the possibility of residual confounding, particularly given the potential for healthy vaccinee bias in observational studies of vaccination.We did not find evidence of an increased risk of preterm birth, small for gestational age at birth, or stillbirth after covid-19 vaccination during any trimester of pregnancy in this large population based study including more than 43 000 births to individuals vaccinated during pregnancy. Our findings—along with extant evidence that vaccination during pregnancy is effective against covid-19 for pregnant individuals and their newborns, and that covid-19 during pregnancy is associated with increased risks of adverse maternal, fetal, and neonatal outcomes—can inform evidence based decision making about covid-19 vaccination during pregnancy. Future studies to assess similar outcomes after immunisation with non-mRNA covid-19 vaccine types during pregnancy should be a research priority.SARS-CoV-2 infection during pregnancy is associated with adverse maternal and birth outcomesPregnancy specific safety information about covid-19 vaccination is important for pregnant individuals, healthcare providers, and policy makers to guide decision makingEvidence from large comparative studies about pregnancy outcomes after covid-19 vaccination during pregnancy is limitedNo association was found between immunisation with an mRNA covid-19 vaccine during pregnancy and increased risk of preterm birth, spontaneous preterm birth, very preterm birth, small for gestational age at birth, or stillbirthThese findings can help inform evidence based decision making about the risks and benefits of covid-19 vaccination during pregnancy


Diffusion tensor imaging (DTI) can assess the structural integrity of the corticospinal tract (CST) in vivo. We aimed to investigate whether CST DTI metrics after intracerebral haemorrhage (ICH) are associated with 6-month functional outcome and can improve the predictive performance of the existing ICH score.We retrospectively included 42 patients with DTI performed within 5 days after deep supratentorial spontaneous ICH. Ipsilesional-to-contralesional ratios were calculated for fractional anisotropy (rFA) and mean diffusivity (rMD) in the pontine segment (PS) of the CST. We determined the most predictive variables for poor 6-month functional outcome [modified Rankin Scale (mRS) > 2] using the least absolute shrinkage and selection operator (LASSO) method. We calculated discrimination using optimism-adjusted estimation of the area under the curve (AUC).p =  0.04) and higher rMD . Discrimination (AUC) values were: 0.76 (95% CI 0.61–0.91) for the ICH score, 0.71 (95% CI 0.54–0.89) for rFA, and 0.72 (95% CI 0.61–0.91) for rMD. Combined models with DTI and non-DTI variables offer an improvement in discrimination: for the best model, the AUC was 0.82 .Patients with 6-month mRS > 2 had lower rFA (0.945 [± 0.139] vs 1.045 [± 0.130]; OR 0.004 [95% CI 0.00–0.77]; In our exploratory study, PS-CST rFA and rMD had comparable predictive ability to the ICH score for 6-month functional outcome. Adding DTI metrics to clinical-radiological scores might improve discrimination, but this needs to be investigated in larger studies. The ICH score has modest discrimination performance for functional outcome , as do other similar prognostic instruments ). This exploratory study suggests that a model with DTI and non-DTI variables might offer an improvement in the prediction of 6-month functional outcome. However, a larger study is required to investigate this further and determine whether any improvement is statistically significant.Our study confirmed the feasibility of using DTI in the acute phase of ICH to extract metrics that can quantify altered microstructure in CST, which is associated with 6-month functional outcome. Predictive performances of DTI metrics alone were comparable to the existing ICH prognostic score ; moreover, the combination of the DTI metrics with the ICH score did not improve the prognostication of outcome. Our findings appear more promising regarding the potential predictive clinical value of DTI metrics; our estimates suggest that combined models with DTI and non-DTI variables (including ICH score) might provide better outcome discrimination than ICH score alone, although the difference in discrimination was not statistically significant. The discrepancy might be due to differences in population characteristics: 68.8% of patients included in the cited study had lobar ICH, and given their variable location and impact on CST fibres, lobar ICH is likely to have less consistent impact on CST. By contrast, we included only patients with deep supratentorial ICH; as shown in Fig. Only one study has compared the predictive performance of DTI parameters and the ICH score  and founThe timing of measuring microstructural changes in DTI is of potential importance because Wallerian degeneration of CST following acute insult is a dynamic evolving phenomenon.  A recentOur findings suggest that quantitative MRI (DTI) is feasible and of potential clinical predictive value that could add information beyond that from standard neuroimaging in acute ICH. Direct anatomical involvement of the CST due to haematoma mass effect or destruction is associated with functional outcome after ICH , which cThe brain MRIs in our study were performed as part of standard clinical acquisition protocols and did not lengthen or reduce tolerability of the examination. Brain MRI already plays a key role in ICH-related diagnostic work-up in practice; it is widely performed to seek evidence of underlying structural causes  and is more sensitive than CT for visualizing biomarkers of the cerebral small vessel diseases that cause most ICH . The potential predictive ability of DTI metrics, in combination with standard clinical and radiological variables, may further consolidate the role of MRI in acute ICH.The modified Rankin Scale is the most widely functional outcome scale adopted in stroke studies but is highly oriented on motor outcome which is highly dependent on CST integrity. However, the clinical impact of ICH goes far beyond motor outcome, including cognitive deficits and dementia , 26. SimOur study has strengths. Our population is homogeneous with respect to the timing of MRI, site of ICH and standardized DTI region of interest (PS-CST). A single-rater assessment of ICH and PHE lesions reduced variability in the segmentation process. We also applied a standardized automated post-processing protocol. In view of the modest sample size, we used LASSO estimation, a technique that produces better models for prediction in small datasets.We also acknowledge limitations. The small cohort size limits the precision of our risk estimates with wide 95% CI, but to the best of our knowledge this is the largest study of deep supratentorial ICH in a Western population. We do not have detailed information on the pattern or severity of motor deficit on hospital admission; however, since all participants had deep ICH and were admitted with clinical symptoms suggesting stroke, it is highly likely that a motor deficit was present in the great majority of those included. A potential advantage of our study is that the findings are generalizable to patients regardless of the severity of their initial motor deficit, which can in any case fluctuate markedly in the first few days after ICH. Although MRI studies are routinely performed for patients with ICH and no contraindications in our centre, the requirement for MRI is still likely to have created a selection bias: the study population includes clinically milder ICH with small hematoma volumes (only 2 patients with baseline GCS < 15 and median ICH volume 5.4 ml [IQR 3.0–11.7]). These aspects might limit the generalizability of our findings to patients with more severe ICH. Nevertheless, the AUC we found for the ICH score is similar to those from studies that evaluated 6-month functional outcome in heterogeneous ICH populations (pooled AUC from four studies: 0.78 [95% CI 0.74–0.82])[In the acute phase after deep supratentorial spontaneous ICH, DTI is feasible as part of routine clinical care; rFA and rMD measured in the normal-appearing pontine segment of the corticospinal tract demonstrated acceptable prognostic ability for 6-month functional outcome. Our findings suggest that DTI measures show promise to improve early prognostication after acute ICH, but further studies in larger cohorts are needed.


Long-acting injectable (LAI) antipsychotics are related to proven compliance to treatment and more constant medication levels (hence the apparent lower side-effect burden).To highlight the experience with LAI antipsychotic treatment during pregnancy.Literature review.Seven cases are reported. A 35year old with schizophrenia received zuclopenthixole LAI (mostly 200mg/monthly) during both her pregnancies . A 35year old with schizophrenia was under risperidone LAI (25mg/2 weeks) and gave birth to a healthy girl at week 37. Another 35year old (probably with schizophrenia) was on olanzapine LAI (300mg/month during the last quarter of her pregnancy) that led to the birth of a healthy girl at week 40. A 37year old with schizophrenia received paliperidone LAI  and gave birth to a healthy boy at week 39. Paliperidone LAI (50mg/monthly) was the treatment of another 34year old with schizoaffective disorder that gave birth to a healthy boy at week 40, as well as of a 26year old (263mg/3-monthly), mother of a healthy boy as well (born at an unspecified week of pregnancy). Finally, a 43year old with bipolar disorder was on aripiprazole LAI (300mg/monthly) during her pregnancy that led to the birth of a healthy girl at week 40.All pregnant women on LAI antipsychotic treatment gave birth to (apparently) healthy babies. LAI doses were mostly low. Long-term follow-up could clarify eventual delayed aftereffects. Based on the literature, LAI antipsychotic treatment could be considered as an option for selected pregnant patients.No significant relationships.