However, selective mortality of more vulnerable children at very high levels of adversity in early life can leave the surviving population to appear on average healthier. This paper estimated the nonlinear effects of postneonatal mortality rate—a proxy for adversity, particularly infections and undernutrition—in a 50 km radius, occurring over the period of infancy, on the subsequent height-for-age and school attendance of the surviving children. The results indicated that an adverse environment in infancy negatively affected height-for-age at age 1–4 years: At relatively low levels of adversity , an additional postneonatal death per 100 person-years decreased height-for-age of the survivors by almost 2% of the mean deficit in height (relative to a common growth standard) when comparing siblings born into different levels of adversity. At high levels of adversity, no effect was found for height-for-age while a small positive association was observed for school attendance at age 7–16 years. The results indicated that selective mortality may have canceled out (or even dominated in the case of school attendance) observable scarring effects following high levels of postneonatal mortality rate in sub-Saharan Africa.Infants in sub-Saharan Africa face adversity: Infections and undernutrition are major causes of infant deaths and can cause physiological damage with long-lasting adverse  •Adversity in early life negatively affects human development in sub-Saharan Africa.•The postneonatal mortality rate over infancy was used as a proxy for adversity.•Elevated adversity in infancy negatively affected height-for-age at age 1–4 years.•At high adversity, selective mortality may cancel out observable scarring effects. Such adscarring effects of early life adversity on adult height, earnings, education, and health have been identified using child mortality rates as proxies for adversity , lower respiratory infections (21.1%), malaria (15%), meningitis (4.5%), protein-energy malnutrition (3.8%), and whooping cough 3.8%; . FurtherAlthough deaths are attributed to specific causes, most deaths result from repeated infections and chronic undernutrition, which also operate in synergy: infections inhibit the uptake of nutrients and suppress appetite, and undernutrition increases susceptibility and mortality from infections . EstimatTherefore, the postneonatal mortality rate is expected to mostly reflect undernutrition and infections causing diarrhea, pneumonia, and malaria, which are all significant determinants of stunted growth . Further2.4Child health varies across socioeconomic status in most contexts, whether measured by the mother's education , father'Scarring or selection mechanisms may be greater for males, as males generally have higher mortality and stunting rates, most likely due to biological frailty .33.1The nationally representative Demographic and Health Surveys were used . Women 1The Demographic and Health Surveys used two-stage sampling. Primary sampling units were first sampled from strata of subnational regions separated into urban and rural areas with a probability proportional to size and then 20–30 households randomly drawn from each primary sampling unit . The pri3.2Child height-for-age z-scores represent standard deviations from a median growth trajectory of healthy children according to the 2006 World Health Organization growth standards . TrainedThe analyses of height-for-age were based on 100 Demographic and Health Surveys (96 when using sibling fixed-effects) from 33 countries . In mostAn indicator for whether 7–16-year-olds attended school in the survey year was obtained from the data on household members. Studies have found child health—measured by, for example, height-for-age—to have determined school attendance in Ghana, Tanzania, and Pakistan . The anaSeveral variables had a small association  with being excluded due to missing outcomes: for example, maternal education, which was slightly lower among children with missing information .3.3All birth histories that included coordinates for neighborhoods were pooled, and sums of person-time and deaths used to construct the postneonatal mortality rate did not necessarily occur within the same survey (but were restricted within countries). Surveys that did not record the outcome variables may still have been used to calculate the postneonatal mortality rate. Age 12 months was classified as postneonatal when constructing the mortality rate, due to age heaping on age at death. Distance between neighborhoods was determined using the haversine formula.pnmnb) was calculated for each target neighborhood n and each target month b as:The rate  and neighborhood p; with neighborhoods p to P being within 50 km of the target neighborhood n (and including n). m and neighborhood p. The numerator was the number of postneonatal deaths  occurring in all neighborhoods p within a 50-km radius from the target neighborhood n over the period between the target month b and b+11 (corresponding to a period of infancy). The denominator was the total person-months at risk  within a 50-km radius from the target neighborhood over the period b to b+11. The postneonatal mortality rate was multiplied by 1200 so that it indicated the number of deaths for every 100 person-years. The postneonatal mortality rate in each target neighborhood and month was linked to the corresponding neighborhood and month of birth of children with outcome measures in the data.The postneonatal mortality rate was limited to neighborhoods and periods with more than four deaths and 600 person-months (or 50 person-years) of exposure within a 50 km radius to improve precision and reduce noise. A total of 218,182 children with a valid height-for-age were excluded due to this limit , and 2973.4Covariates were child's age (in months), birth order, birth interval , and three binary indicators for whether the child was a firstborn, female, and twin . Maternal education (in years), mother's age at birth (in years), and the number of siblings ever born to mother were also added to models using neighborhood fixed-effects. Squared terms were included for mother's age at birth, since young age can be harmful due to immaturity or low socioeconomic status while advanced age may also be negative for biological reasons ; and chiExcept for the child's age, these covariates were unlikely to reflect confounders. However, not accounting for age could cause effects to be overestimated: older children may have suffered negative exposures over a longer period, which decreased their height-for-age , as well3.5y was one of the two outcomes, x was a vector of covariates, and b and neighborhood n of child i in sibling pair s (or sibling group in a few cases). A squared term for postneonatal mortality rate was included to allow for nonlinearity due to the increasing role of selective mortality at higher adversity levels. Postneonatal mortality rate was centered around percentile 25 since scarring effects were expected at the lower end of the distribution: therefore, Sibling fixed-effects models were estimated to account for all households and sibling-level factors. Additionally, neighborhood fixed-effects models were estimated for comparison. The models were specified as ordinary least squares regression models:e:Testing whether the effects of adversity in infancy varied by socioeconomic status  was explored by interacting postneonatal mortality rate with an indicator for whether the mother of the child had any education For twins, only one twin was considered  for analyses using sibling fixed-effects since the postneonatal mortality rate did not vary between twins. For height-for-age, 141,603 children did not have a sibling with valid data and were excluded from siblings fixed-effects models, and 2874 neighborhoods had only a single valid observation and were excluded from neighborhood fixed-effects models . For schFirst, results from regression models were tabulated. Second, the marginal effects of the postneonatal mortality rate were estimated by first taking the partial derivative of the regression equation with respect to postneonatal mortality rate, and then estimating across the distribution of postneonatal mortality rate. The resulting marginal effects were graphed between percentile 5 and 95 of the postneonatal mortality rate distribution and then tabulated at percentiles 1, 5, 10, 25, 50, 75, 90, 95, and 99. The results from neighborhood and sibling fixed-effects models were expected to be similar since it is unclear how sibling-level factors influence postneonatal mortality rates. Results from models using survey-fixed effects were shown in the Supplement . The exp3.6Ten sensitivity analyses were done: 1) Deaths and children at risk were further restricted to a 25 km radius when constructing postneonatal mortality rate . 2) Post4) Neonates were included in the adversity measures, which was, therefore, an infant mortality rate : Since n5) Cubed term for postneonatal mortality rate was added to the regression models to test the robustness of the results to a more irregular relationship . 6) Birt7) Neighborhood fixed-effects models were estimated on samples restricted to children that had valid sibling data to explore whether differences in estimates between neighborhood and sibling fixed-effects models were due to sample restrictions . 8) Post44.1On average, children were 1.8 z-score  below the median in the reference population and 76%  attended school . Postneo4.2A single postneonatal death increase per 100 person-years was associated with a 0.015  z-score lower height-for-age  when estimating neighborhood fixed-effects models . When usWhen using neighborhood fixed-effects, the association between postneonatal mortality rate and school attendance indicated 0.0016  lower probability of attending school for an additional death at percentile 25 of postneonatal mortality rate. When using sibling fixed-effects models, that association was small and not statistically significant: However, the coefficient for the squared term of postneonatal mortality rate was positive and statistically significant. Females had a 0.028  lower probability of attending school.The marginal effect of postneonatal mortality rate on height-for-age was negative at the bottom of the postneonatal mortality rate distribution, while it attenuated at higher rates of postneonatal mortality, eventually reaching zero at the top of the distribution . For schThe interquartile range of postneonatal mortality rate was 4.2 deaths . At the At the bottom of the postneonatal mortality rate distribution, at percentile 10, an additional postneonatal death per 100 was associated with a 0.0024  decrease in the probability of attending school when using neighborhood fixed-effects models and a non-statistically significant 0.0013  decrease when applying sibling fixed-effects models. At the top of the postneonatal mortality rate distribution, for example at percentile 90, an increase of a single postneonatal death per 100 was associated with a 0.0028  increase in the probability of attending school when using neighborhood fixed-effects and a 0.0029  increase when using sibling fixed-effects.Survey fixed-effects models indicated a negative effect of postneonatal mortality rate on both height-for-age and school attendance across the whole distribution of postneonatal mortality rate .4.3When using neighborhood fixed-effects models, the marginal effects of postneonatal mortality rate on height-for-age for children of mothers with education overlapped with that of children of mothers without education across the postneonatal mortality rate distribution . When usWhen using neighborhood fixed-effects, the marginal effect of postneonatal mortality rate on height-for-age was considerably stronger for females, particularly the negative association observed at the bottom of the postneonatal mortality rate distribution . The intFor boys, the marginal effects of postneonatal mortality rate on school attendance were positive across the whole distribution of postneonatal mortality rate and increased somewhat as the postneonatal mortality rate increased . For females, the marginal effect of postneonatal mortality rate on school attendance was negative at the bottom of the distribution of postneonatal mortality rate, decreased at higher rates of postneonatal mortality, and was positive at high levels of postneonatal mortality rate. Females had a 0.011  lower probability of attending school  when estimated using sibling fixed-effects .4.4Sensitivity analyses were broadly consistent with the main results: some statistically significant coefficients in the main analyses were not in the sensitivity analyses, although remaining similar in magnitude.As expected, replacing the postneonatal mortality rate with infant mortality rate indicated much smaller effect sizes, particularly for height-for-age, likely due to less weight on undernutrition and infections in that measure .Adding cubed postneonatal mortality rate to the regression equation indicated that a regression with just a linear and squared term approximated the relationships similarly : HoweverIndexing postneonatal mortality rate to the 12-month period starting at conception yielded mostly null results .Results for school attendance were more similar between neighborhood and sibling fixed-effects models after converting postneonatal mortality rate to natural log scale .55.1The findings suggest a negative effect of exposures to adversity in infancy on height-for-age at age 1–4 years, indicating net-scarring effects on human development for those who survive. These effects were strong at the bottom of the distribution of postneonatal mortality rate: For example, at percentile 10 , a 0.03 decline in height-for-age was observed for a further increase of a single postneonatal death per 100 person-years between siblings in infancy—or 1.7% of the sample mean deficit in height-for-age relative to the 2006 World Health Organization growth standard. This effect size is rather small but meaningful considering that the interquartile range of postneonatal mortality rate was 4.2 deaths. To put this effect size into perspective: complementary feeding interventions have resulted in 0.11–0.25 z-score increase in length-for-age  and elimFor school attendance, the findings only indicated a negative net-scarring effect at the bottom of the distribution of postneonatal mortality rate and only when comparing children within the same neighborhood and not when comparing siblings. The marginal effects of postneonatal mortality rate on school attendance were positive but small above the median postneonatal mortality rate, and similar between neighborhood and sibling fixed-effects models at high rates: For example, at percentile 90, a further increase of a single postneonatal death per 100 person-years between siblings in infancy was associated with a 0.0029 increase in the probability of attending school at age 7–16 years—or about 0.4% of the sample mean school attendance.The absence of a marginal effect between postneonatal mortality rate and height-for-age at rates of postneonatal mortality that were already relatively high was consistent with the selection mechanism increasing at high levels of adversity . In thisChildren of mothers without education had a stronger association between postneonatal mortality rate and height-for-age, and the marginal effect was close to zero across the distribution for children of mothers with education: However, this heterogeneity was only observed when comparing siblings, where all family level factors were taken into account.The marginal effect on height-for-age was somewhat stronger for girls than boys, particularly at low rates of postneonatal mortality, which may either indicate that girls were more vulnerable to exposures to adversity or that selective mortality cancels out more of the net-scarring effects on boys. The latter interpretation is more consistent with the observation that boys tend to be more sensitive to adverse exposures due to biological frailty . HoweverThere was major heterogeneity in the effects of postneonatal mortality rate on school attendance by sex. The effect was positive for boys across the postneonatal mortality rate distribution. For girls, it was negative at the lower end of the distribution and positive at the top of the distribution.  Applying the interpretation that selection effects were greater among boys, the positive link between school attendance and postneonatal mortality rate in infancy would suggest that selection effects dominate over any scarring effects, even at relatively low levels of adversity. The increase in the positive marginal effect on school attendance at higher rates of postneonatal mortality was modest for boys, which would suggest that the increase in selection effects at greater levels of adversity was also modest. For girls, the results were consistent with scarring effects dominating over selection effects at low levels of adversity, while selection effects were more important at high levels of adversity.However, acquired immunity, reduced cohort sizes due to mortality, and increased investments in surviving children or other factors cannot be ruled out as mechanisms causing higher school attendance among those exposed to a higher postneonatal mortality rate in infancy. For example, sons might have received greater investment following times of elevated adversity. In addition, effect sizes were small. Factors such as acquired immunity and greater investments in surviving children following adversity may also be an alternative to selective mortality biasing the effects on height-for-age downward.Results were mostly consistent between neighborhood and sibling fixed-effects models. Still, differences may be related to omitted family level variables biasing the estimates from the neighborhood-fixed effects models . The differences in results obtained from survey fixed-effects models using spatial and temporal variation were consistent with correlated environments and a more general level of harmful exposures in areas where the average rate of postneonatal mortality was greater, exaggerating the negative effects of exposures to adversity when not accounting for the spatial variation in some form .5.2A limitation of this paper was the use of a general measure of adversity that captures a host of infectious diseases in interaction with undernutrition. Other harmful exposures such as air pollution or armed conflict cannot be ruled out as contributing to adversity, although, these factors also affect child mortality through undernutrition and infections . FurtherCalculating a relatively high-resolution estimate, such as the postneonatal mortality rate in this paper, requires enough deaths and exposure time to acquire a reliable estimate. Therefore, these estimates were not representative, as areas that were more extensively surveyed, more densely populated, and had higher mortality were more likely to be included in the analyses. These areas were less likely to be rural but also tended to be more disadvantaged overall, when measured as maternal education (but more advantaged when comparisons were made within each survey).Although the exposure used in this paper was exogenous with respect to child health, its effects were strictly not causal due to misclassification and noise in the adversity measure: First, not all children faced adversity when the postneonatal mortality rate was high, and some children faced adversity even when the postneonatal mortality rate was low, which has similar implications as misclassification . Also, hFurther, fixed-effects models can enhance the problem of misclassification, or measurement errors, most likely biasing estimated effects sizes toward zero . TherefoFinally, selection effects may have biased the observed net-scarring effects downward, even at relatively low rates of postneonatal mortality, which was still high in the context studied here .6In sub-Saharan Africa, being exposed to elevated postneonatal mortality rate during infancy negatively affected the observed height-for-age of surviving children at age 1–4 years. However, for increases at rates of postneonatal mortality that were already very high, selective mortality may have canceled out net-scarring effects. School attendance at ages 7–16 was slightly greater for children exposed to increased adversity during infancy when postneonatal mortality rate was already high. Selective mortality may obscure some of the detrimental consequences of adversity in early life when studying the outcomes of the survivors. However, although effect sizes were rather small, particularly for school attendance, they were likely to be underestimated.This project used publicly accessible secondary data obtained from the Demographic and Health Surveys (DHS) website. The DHS data were not collected specifically for this study and the author had no access to identifiers linked to the data. These activities did not meet the regulatory definition of human subject research. Therefore, an Institutional Review Board (IRB) review was not required. The Harvard Logwood Campus IRB allows researchers to self-determine when their research does not meet the requirements for IRB oversight via guidance online. The ICF IRB and local IRB approved data collection procedures and questionnaires and the U.S. Center for Disease Control and Prevention (CDC) reviewed protocols.Karlsson was the sole author of this study.None.https://dhsprogram.com (requiring a simple application).The Demographic and Health Surveys are available at None.


Among the three vaccines for which protection was evaluable (Moderna (mRNA-1273); Sinopharm (Vero Cell-Inactivated); Serum Institute of India (ChAdOx1 nCoV-19)), only the Moderna vaccine was associated with significant protection . Protection by the receipt of any vaccine against severe disease was 85% , with protection estimates of 75% to 100% for the three vaccines. Conclusions: Vaccine protection against COVID-19 disease of any severity caused by the Delta variant was modest in magnitude and significant for only one of the three evaluable vaccines. In contrast, protection against severe disease was high in magnitude and consistent for all three vaccines. Because our findings are not in complete accord with evaluations of the same vaccines in more affluent settings, our study underscores the need for country-level COVID-19 vaccine evaluations in developing countries.Background: From May to December 2021, Bangladesh experienced a major surge in the Delta variant of SARS-CoV-2. The earlier rollout of several vaccines offered the opportunity to evaluate vaccine effectiveness against this variant. Methods: A prospective, test-negative case-control study was conducted in five large hospitals in Dhaka between September and December 2021. The subjects were patients of at least 18 years of age who presented themselves for care, suffering COVID-like symptoms of less than 10 days’ duration. The cases had PCR-confirmed infections with SARS-CoV-2, and up to 4 PCR test-negative controls were matched to each case, according to hospital, date of presentation, and age. Vaccine protection was assessed as being the association between the receipt of a complete course of vaccine and the occurrence of SARS-CoV-2 disease, with symptoms beginning at least 14 days after the final vaccine dose. Results: In total, 313 cases were matched to 1196 controls. The genotyping of case isolates revealed 99.6% to be the Delta variant. Receipt of any vaccine was associated with 12% (95% CI: −21 to 37,  The ongoing COVID-19 (coronavirus disease 2019) pandemic caused by severe acute respiratory syndrome coronavirus 2 (SARS-CoV-2) has been responsible for more than 416 million cases and around 6 million deaths to date ,4, as haEffectiveness studies, performed after deployment, have been critical to monitoring the protection offered by vaccines as the pandemic has progressed, especially because vaccines have been challenged by many of the new variants that have arisen . HoweverThe study was carried out in four designated COVID-19 hospitals  in Dhaka, Bangladesh, between 8 September 2021 and 29 December 2021. For this study, we prospectively enrolled patients who presented with COVID-19-like symptoms for RT-PCR testing, to identify the test-positive cases and test-negative controls.Patients attending the outpatient isolation ward or inpatient unit of the designated hospitals with symptoms of SARS-CoV-2-like illness  and having performed an RT-PCR test for COVID-19 were approached for enrollment in the study. Patients aged ≥ 18 years, the age group targeted for vaccination before and during the study period, were considered to be eligible for the study if they had not been previously diagnosed with COVID-19 using RT-PCR testing and if the duration of symptoms before presentation was ≤10 days. Enrollment of patients was carried out in all health facilities from 8:30 to 14:00, six days a week, and informed written consent was obtained before enrollment in the study.Upon presentation for testing, patients were approached and checked for eligibility criteria by a study nurse, and informed written consent was obtained. Each patient’s clinical information, including history and physical examination, was entered into structured data forms by a study physician. Disease at the time of presentation was classified as severe or not severe, using the WHO criteria . SpecialAt the time of presentation, the earlier receipt of the vaccine was ascertained by inspection of the official vaccination cards provided by the Government of Bangladesh (GoB) during vaccination. For persons not having cards, but giving a verbal history of vaccination, vaccination was confirmed by linking the national identification number of the subject to the centralized database of vaccination records of the management information system (MIS) of the GoB. Information was obtained about the date and brand of each dose. All ascertainment and linkage of vaccine histories and records was performed in a manner blinded to the patient’s RT-PCR results, and, thus, to the patient’s status as a case or a control. Complete vaccination was defined as the receipt of two doses of the same vaccine; subjects who had received heterologous two-dose regimens or who received the vaccine outside of Bangladesh were excluded.At the time of presentation, a nasopharyngeal swab (NPS) specimen was collected from both nares using a swab, and the swab was then placed in a viral transport medium (VTM). The VTM, along with the swab stick, was placed in a cooler box with an ice pack (maintaining 2 to 8 °C) and transported to the Virology Laboratory of the International Centre for Diarrhoeal Disease Research, Bangladesh  for specimens from Dhaka Medical College and Hospital, Kurmitola General Hospital, and Mugda Medical College and Hospital, and sent to the Institute of Epidemiology and Disease Control Research (IEDCR) laboratory in the case of specimens collected from Bangabandhu Sheikh Mujib Medical University Hospital.github.com/cov-lineages/pangolin; accessed on 15 January 2022) classification system.Viral RNA from NPS was extracted using the Qiagen Viral RNA mini-kit , following the manufacturer’s instructions. Real-time PCR (RT-PCR) was performed by using CDC_N2, RdRp, and ORF3a primers and probes with the iTaq Universal Probes Reaction Master Mix , iScript Reverse Transcriptase. The RNaseP housekeeping-gene RNA level was performed to determine the quality of the NPS sample. A positive RT-PCR result with a ≤40-cycle threshold (CT) value was taken as positive. Other results were classified as negative. Once the RT-PCR test result was available, the laboratory result was entered digitally into a special form. RT-PCR-positive samples with a ≤30 CT value were also subjected to whole-genome sequencing using the MinION sequencing platform, carried out at the Institute for Developing Science and Health Initiatives (ideSHi) laboratory. Lineages were assigned by the SARS-CoV-2 pangolin  and controls (RT-PCR-negative patients) were listed for each day, and controls were selected for each case at each hospital at a ratio of up to 4:1 by a statistician blinded to the vaccination status of the subjects. Controls were matched to cases by site and date of presentation for care and age on the date of testing  and were randomly sampled if the number of eligible, matched control patients exceeded 4 for a given case. If 4 matched controls could not be found for a case on the date of presentation, the sampling frame was progressively widened, beginning with the day before, then the day after, and so on until a frame of ±3 days was reached. The selection of cases followed the principles of incidence density sampling: cases were selected in sequential order and could not be later resampled as cases, but controls could later be resampled as cases or controls.For the purposes of this analysis, we incorporated data on mortality from a follow-up conducted 30 days after enrollment.p < 0.05 (two-tailed) with at least 80% power, assuming that 12% of controls were vaccinated, with 4 controls per case. These assumptions yielded a required size of 323 cases and 1292 controls. Our primary analysis examined the protection given by a complete two-dose series of any homologous regimen of vaccines administered by the GoB at least 14 days before presentation for symptomatic episodes of RT-PCR-confirmed SARS-CoV-2 disease of any level of severity. In secondary analyses, also pre-specified in the statistical analysis plan, we evaluated the protection offered by complete vaccine regimens against severe SARS-CoV-2 disease, as well as the protection offered by a complete regimen by duration since the receipt of the second dose. Severe disease in these analyses was defined as “presenting severe disease” at the time of presentation for testing, using the WHO criteria [Our primary goal, formally documented in a statistical analysis plan that was finalized before analyzing the data, was to assess the overall impact of all vaccines rolled out in Bangladesh upon all episodes of detected COVID disease. Accordingly, our sample-size calculation was designed to detect 60% protection against symptomatic SARS-CoV-2 disease by the receipt of a complete regimen of any COVID-19 vaccine at t-test for the dimensional variables and used unconditional logistic regression models to estimate vaccine protection. The odds ratios for the associations between vaccination and disease status were estimated by exponentiation of the coefficient for vaccination from the fitted models, and the standard error of the coefficient was used to calculate 95% confidence intervals for the estimated effect. Protective effectiveness was calculated as ((1 − odds ratio) × 100%)). For evaluation of vaccine protection against all disease episodes, severity at presentation was forced into the models as an independent variable, in order to help protect against healthcare utilization bias [p-values ≤ 0.10 (two-tailed), and if the associations remained significant at p < 0.05 (two-tailed) in the stepwise variable selection models. This threshold was also applied when judging the significance of protective associations between vaccination and disease status.Bivariate analyses were performed in a matched fashion and conducted between relevant demographic, behavioral, and clinical variables and disease status using bivariate conditional logistic regression models. For the evaluation of vaccine protection, we used conditional logistic regression, considering the matched case-control status as the outcome and vaccination status and selected covariates as independent variables, taking the non-receipt of vaccine as the referent category for the assessment of vaccine protection. For unmatched analyses, we compared cases and controls with the chi-square test for the categorical variables and Student’s ion bias . Other pBecause of the paucity of severely ill SARS-CoV-2 patients enrolled in the study, we also conducted an exploratory analysis comparing verified cases presenting severe disease versus controls presenting severe disease, disregarding the matched selection. The restriction of cases and controls to those presenting severe disease was employed to help further safeguard against residual healthcare utilization bias, as both the uptake of vaccination and the use of healthcare facilities may differ in those presenting with severe versus non-severe disease. In this analysis, unconditional logistic regression models were fitted, and the matching variables for the site of presentation, time of presentation (the month was used), and age at presentation were forced into the models as covariates. All analyses were performed after a formal data lock and according to the analyses outlined in a statistical analysis plan. We used R statistical software (version 4.10) for data analyses. The clogit package of R was used to estimate the conditional logistic regression coefficients by maximizing the conditional likelihood.Before initiation of the study, ethical approval from the Research and Ethical Review Committees of the icddr,b was obtained. Informed written consent was obtained from the participants prior to enrollment in the study.We approached 14,126 patients attending the hospitals in question for COVID-19 testing. Of these, 3291 eligible patients were enrolled after giving their informed consent. Among these subjects, we identified 362 RT-PCR-positive cases and were able to select 1416 matched RT-PCR-negative controls. Finally, 313 cases and 1196 matched controls were included in the analysis, after restricting the inclusion of vaccinated subjects to those who had received an adequately documented, complete, homologous vaccine regimen in Bangladesh at least 14 days before presentation for testing . No subjIn simple comparisons with matched cases and controls , we founp = 0.423) (p = 0.019) (p = 0.079) (The adjusted protective effectiveness (PE) value against all episodes of COVID-19 disease conferred by the receipt of a complete regimen of any vaccine rolled out was 12%  . For COV= 0.019) . Results= 0.079) .p = 0.013) of protection by any vaccine against symptomatic COVID-19 disease in those participants who were enrolled within 19 weeks of the second dose (p = 0.182).To evaluate vaccine protection by time since the receipt of the second dose, we approximately bisected the period of enrollment into the period of ≤19 weeks versus the period of >19 weeks after the second dose and retained the matched selection strategy. Most positive cases and controls were vaccinated with the Serum Institute of India (ChAdOx1 nCoV-19) vaccine and were enrolled more than 19 weeks before presentation; most cases and controls were vaccinated with either the Moderna (mRNA-1273) or Sinopharm (Vero Cell-inactivated) vaccine and were enrolled within 19 weeks before presentation . We founond dose . Beyond p = 0.029). While there was a suggestion of protection by the Sinopharm (Vero Cell-Inactivated) vaccine , the point estimate for the Serum Institute of India vaccine (ChAdOx1 nCoV-19) revealed no protection   vaccine showed significant protection against all episodes of COVID-19 disease  .p = 0.098) and Moderna (mRNA-1273) vaccines , but not for the Serum Institute of India vaccine . However, wide 95% confidence intervals again limited our interpretation of these estimates  protection, the Sinopharm (Vero Cell-Inactivated) vaccine, 34% protection  and the Serum Institute of India (ChAdOx1 nCoV-19) vaccine, 3% protection  during the first 19 weeks. Beyond 19 weeks after the second dose, there was no subject who had been vaccinated with the Moderna (mRNA 1273) vaccine, and no protection was observed for the Serum Institute of India (ChAdOx1 nCoV-19) or the Sinopharm (Vero Cell-Inactivated) vaccines  vaccine exhibited 64% , but that protection disappeared in the subsequent 19 weeks .In contrast, protection by any deployed vaccine against ultimately severe COVID-19 disease over the entire study interval was of high grade , a result that was corroborated when cases and controls were limited to patients who presented with severe disease, a tactic designed to help remove the residual bias due to differential health-seeking behavior in cases versus controls.Because protection by different COVID-19 vaccines has been shown to differ in past studies, we conducted exploratory analyses of these relationships for the different vaccines deployed. Over the entire study interval, only the Moderna (mRNA-1273) vaccine was associated with significant protection against all COVID-19 disease episodes, albeit of moderate magnitude . Conversely, the Serum Institute of India (ChAdOx1 nCov-19) vaccine failed to exhibit any protection. A wide confidence interval precluded inferences about protection by the Sinopharm (Vero-Cell Inactivated) vaccine. Although confidence intervals were wide, point estimates of protection against ultimate severe COVID-19 disease were high for all three vaccines (75% to 100%).p-values. Nonetheless, our exploratory analyses did reveal statistically significant protection by the Moderna vaccine against disease of any severity, and high-grade, statistically significant protection by the receipt of any vaccine against severe disease, with consistently high point estimates of protection for each of the three vaccines in predominant use. Finally, although the Pfizer-BioNtech (BNT162b2 mRNA) vaccine and the Sinovac (Vero Cell-Inactivated) vaccine have been introduced into Bangladesh, their use was not common enough for exploratory analyses of protection by each vaccine.It is important to consider several potential limitations of our study. As a study based in just five hospitals serving an urban population of nearly 20 million people who have many other sources of care available, our findings cannot be taken as a basis to generalize for the entire city of Dhaka. However, the use of the sampling frame provided by the test-negative design allowed the selection of controls who could reasonably be considered to represent persons who would have been selected as cases had they developed and sought care for COVID-19. This feature, in conjunction with exhaustive analytical control for potential confounding variables, prospective conduct of the study, determination of eligibility, and acquisition of informed consent without knowledge of the PCR status of subjects, and the ascertainment of vaccination status based on objectively verified sources, obtained in a fashion without knowledge of subjects’ status as cases or controls, likely ensured a high level of internal study validity. Moreover, extensive genotyping of SARS-CoV-2 isolates from the cases in our study allowed us to conclude that our findings pertain to the Delta variant, which was responsible for disease in virtually all cases. Another limitation of our study was that it was powered for the assessment of receipt of any vaccine, rather than of individual vaccines, and was not well-powered to evaluate protection against clinically severe SARS-CoV-2 disease, accounting for a wide 95% confidence interval and high Our estimates of vaccine protection against episodes of SARS-CoV-2 disease of any severity were generally lower than those reported elsewhere for the Delta variant. We found 64% protection from the Moderna (mRNA-1273) vaccine against symptomatic SARS-CoV-2 infection, whereas, in a study conducted in the United States, the vaccine showed 93% protection against symptomatic disease caused by the Delta variant . A test-In contrast, our estimates of vaccine protection against severe SARS-CoV-2 disease are largely consistent with estimates of vaccine protection against the Delta variant reported elsewhere. Our evaluation found a point estimate of 75% protection by the Sinopharm (Vero Cell-Inactivated) vaccine against ultimate severe disease; the vaccine conferred 95% protection against hospitalization and critical admissions caused by the Delta variant in the United Arab Emirates . In IndiOur study is the first to evaluate the protective effectiveness of the SARS-CoV-2 vaccines deployed in Bangladesh. The fact that our study, which used the same test-negative design approach widely employed for real-world evaluations of COVID-19 vaccines worldwide, revealed lower estimates of vaccine protection against COVID-19 diseases of any severity but showed similar protection against severe disease in comparison with studies of the same vaccines conducted in other countries, may in part reflect the wide confidence intervals of our estimates for the exploratory analyses, but may also illustrate that the levels of vaccine protection against the Delta variant may vary according to the geographical context of the evaluation. These findings underscore the importance of the continued country-level monitoring of vaccine effectiveness as the SARS-CoV-2 pandemic evolves and will inform future models of SARS-CoV-2 vaccine deployments. The Delta surge in Bangladesh has now passed, and we are continuing the study during the current surge by the Omicron variant, which likely will offer additional insights.


Are there patient-level or systemic racial disparities in meeting fracture care time-to-surgery benchmarks?This cohort study was a secondary analysis of prospectively collected multicenter data for 2565 patients with hip and femur fractures enrolled in 2 randomized trials in the US and Canada. After controlling for patient-level characteristics, there was an association between missing the 24-hour benchmark for surgery and the proportion of insured patients cared for by the hospital, and a separate association between proportion of insured patients and the racial and ethnic minority patients cared for by hospitals.These results suggest that patients seeking care from institutions with more diverse and more uninsured patients are more likely to experience delays greater than the 24-hour benchmarks, regardless of the individual patient race, implying structural health systems biases. This cohort study examines the association of race, ethnicity, and insurance coverage status at the individual and hospital level with time to surgery among patients with hip and femur fractures. Racial disparities in treatment benchmarks have been documented among older patients with hip fractures. However, these studies were limited to patient-level evaluations.To assess whether disparities in meeting fracture care time-to-surgery benchmarks exist at the patient level or at the hospital or institutional level using high-quality multicenter prospectively collected data; the study hypothesis was that disparities at the hospital-level reflecting structural health systems issues would be detected.This cohort study was a secondary analysis of prospectively collected data in the PREP-IT  program from 23 sites throughout North America. The PREP-IT trials enrolled patients from 2018 to 2021, and patients were followed for 1-year. All patients with hip and femur fractures enrolled in the PREP-IT program were included in analysis. Data were analyzed April to September 2022.Patient-level and hospital-level race, ethnicity, and insurance status.Primary outcome measure was time to surgery based on 24-hour time-to-surgery benchmarks. Multilevel multivariate regression models were used to evaluate the association of race, ethnicity, and insurance status with time to surgery. The reported odds ratios (ORs) were per 10% change in insurance coverage or racial composition at the hospital level.P = .005) and the interaction term between hospital population insurance coverage and racial composition . There was no association between patient race and delay beyond 24-hour benchmarks .A total of 2565 patients with a mean (SD) age of 64.5 (20.4) years  were included in analysis. Of these patients, 834 (32.5%) were employed and 2367 (92.2%) had insurance; 1015 (39.6%) had sustained a femur fracture, with a mean (SD) injury severity score of 10.4 (5.8). Five hundred ninety-six patients (23.2%) did not meet the 24-hour time-to-operating-room benchmark. After controlling for patient-level characteristics, there was an independent association between missing the 24-hour benchmark and hospital population insurance coverage (OR, 0.94; 95% CI, 0.89-0.98; In this cohort study, patients who sought care from an institution with a greater proportion of patients with racial or ethnic minority status or who were uninsured were more likely to experience delays greater than the 24-hour benchmarks regardless of the individual patient race; institutions that treat a less diverse patient population appeared to be more resilient to the mix of insurance status in their patient population and were more likely to meet time-to-surgery benchmarks, regardless of patient insurance status or population-based insurance mix. While it is unsurprising that increased delays were associated with underfunded institutions, the association between institutional-level racial disparity and surgical delays implies structural health systems bias. To this end, the aims of this study were to evaluate whether (1) there are racial disparities in meeting time-to-operating room benchmarks for time-sensitive hip and femur fractures using high-quality prospectively collected data and (2) delays were attributable to patient factors  or institutional factors . A 24-hour time-to-operating room benchmark is well-accepted for both femoral shaft and hip fractures to reduce morbidity and mortality.NCT03385304) and PREPARE (Preoperative Alcohol Skin Solutions in Fractured Extremities) (NCT03523962).26 Enrollment took place in 23 sites across the US and Canada. Of these 23 sites, 1 was a military medical center and all others were private nonprofit institutions. Twenty-one sites were designated American College of Surgeons Level I trauma centers and 2 sites were Level II trauma centers. Twenty states or provinces in the US and Canada were represented geographically, with multiple  sites in all US Census regions  and in Canada in both urban and rural locations. The PREP-IT trials enrolled patients from 2018 to 2021 and patients were followed for 1 year. PREP-IT trial participants were aged 18 years or older and presented with either an open upper or lower extremity fracture or a closed fracture of the lower extremity or pelvis. This study followed the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) reporting guideline for cohort studies.This cohort study was a secondary analysis of data collected in the Program of Randomized Trials to Evaluate Preoperative Antiseptic Skin Solutions in Orthopaedic Trauma (PREP-IT) program. The PREP-IT trials were approved by a central institutional review board (Advarra). The PREP-IT program is comprised of 2 parallel cluster randomized crossover trials: Aqueous-PREP (Preoperative Aqueous Antiseptic Skin Solutions in Open Fractures)  or closed hip fracture  were eligible for inclusion. The outcome investigated was time to surgery bifurcated into surgery within 24 hours of hospital admission or greater than 24 hours from hospital admission. No enrolled patients were excluded from the analysis.Patients were dichotomized as White or racial and ethnic minority patients based on self-reported identification within 12 initial race and ethnic categories. Patients not identifying as White were grouped together in order to cluster all underrepresented minorities who may experience disparities relative to their White counterparts, which resulted in a more straightforward regression model that optimized statistical power and thus ability to draw conclusions. In particular, many of the 12 racial and ethnic categories had quite small patient numbers, which would make comparisons impossible. Hospital population racial composition was defined based on hospital racial and ethnic minority patient population. Insurance status was coded as either insured or uninsured. Both hospital population racial composition and hospital population insurance were calculated as the proportion of patients who identified as a racial and ethnic minority and without insurance, respectively. As an example, a hospital site that treated 10 patients without insurance and 90 patients with insurance would result in 10% of the site’s population  being uninsured.t tests for continuous variables and χ2 tests for categorical variables. Considering the nested structure of the data focusing on person-level and institution-level independent variables of the person-level dependent variable, a mixed effects modeling approach was employed for the primary analysis. Toward developing the mixed effects model, we used a model building approach that included stepwise variable block inclusion, largely following the approach outlined, by first modeling steps offered with patient-level variables included and then by introducing institution-level variables .32 Specifically, person-level continuous independent variables were mean-centered. Institution-level independent variables were constructed by determining proportion of patients that met the specific criteria  as compared with total number of patients served by each institution. An interaction term was included between institution-level patient makeup by racial and ethnic minority status and insurance status to understand the combined effect on the focal outcome of a patient’s probability of meeting the 24-hour surgical window. Due to the interaction term being significant (P < .05 in 2-sided tests), it was further probed via a model-implied plot. Additionally, curvilinear treatment of institution-level racial composition of patients served was empirically tested via the inclusion of quadratic and cubic terms, which did not demonstrate significance either comparing omnibus model fit statistics with the more parsimonious model or individual model parameters. The variance inflation factor was evaluated for all model predictors and met acceptable thresholds (below 5)  age 64.5 (20.4) years ; 2112 patients (82.3%) were White  . Of thesP < .001), more likely to be women , less likely to be employed , had a higher American Society of Anesthesiologists physical status classification  and more often suffered a femur fracture as opposed to hip fracture   did not meet 24-hour time to operating room benchmarks. Patients who did not meet the 24-hour standard-of-care surgical window were frequently older  . DistribThere was variability across sites with regards to meeting 24-hour benchmarks and race and ethnicity distribution, as well as population-based health insurance. Institutions met 24-hour time-to-operating-room benchmarks through a range of 45.2% (196 of 433 procedures) to 97.4% (37 of 38 procedures) . MinoritP = .08). However, increasing hospital population insurance coverage was protective against missing time-to-surgery benchmarks . At the patient level, there was no association between missing 24-hour time-to-surgery benchmark and race or ethnicity  and there was no independent association between hospital population racial composition and surgical delay  (P = .03) (P = .49). There were independent associations between missing the 24-hour benchmark and age , American Society of Anesthesiologists class, patient employment status , and patient BMI .After controlling for patient-level characteristics, there was not an independent association between missing 24-hour time-to-surgery benchmark and patient-level insurance status  . There wP = .03) . These rP = .03) . Of noteP = .03) . There wAt low rates of uninsured patients, the probability of missing the 24-hour time-to-surgery benchmark was 12.5% to 14.6% when racial composition varied from 0% to 50% racial and ethnic minority patients . ConversThe results of this study demonstrate disparities in meeting fracture care time-to-operating room benchmarks at the hospital level rather than at the patient level. Patients seeking care from hospitals with more diverse and less insured patient populations were at higher risk of delayed surgical treatment beyond 24-hour time-to-operating room benchmarks for both femur and hip fractures. For example, at low rates of uninsured patients, there was minimal difference between probability of missing the 24-hour time-to-surgery benchmark, but at higher rates of uninsured patients, there were stark differences between institutions with different racial composition. At the low end, when racial composition had fewer racial and ethnic minority patients, the probability of missing the 24-hour window was less than 1%; however, as the racial composition of racial and ethnic minority patients increased, the risk of missing the 24-hour window also increased, reaching almost 18% when racial composition was 50% racial and ethnic minority patients. Hospitals that treated less diverse  patient populations appeared to be more resilient to the mix of insurance status of their patient population, and did not have an association between worsening delay to surgery with increasing uninsured patient population. It is likely that well-funded health systems caring for a higher proportion of insured patients have in place quality improvement programs and other support structures (such as operating room access) that ensure appropriate time-to-operating room for these time-sensitive procedures, and that these programs may be missing at institutions caring for more racially diverse and less insured patient populations. More research is needed to explore health system programs or structures that explain these findings.This analysis also demonstrated that increasing age, increasing medical comorbidities (reflected in ASA class), and increasing BMI were, not surprisingly, associated with increased risk of missing the 24-hour time to surgery benchmark, as these patients likely required more extensive medical optimization. Additionally, employed patients were less likely to miss this benchmarks. These results likely reflect the relative health and vigor of the patient cohort .27 Two prior studies have demonstrated racial disparities at the structural health systems level both in time to surgery as well as hospital length of stay for hip fractures.27 The results of this study are also consistent with several prior studies demonstrating delays in hip fracture treatment among racial and ethnic minority patients28 and worse outcomes among racial and ethnic minority patients.28 As these studies only evaluated disparities on the patient level, it is impossible to know whether their findings were due to patient-specific biases or structural health systems issues. The results of this study were also consistent with a study within the Kaiser Permanente patient population demonstrating that in their integrated managed care system, in which all patients have health insurance coverage, racial and ethnic minority patients with hip fracture had postoperative mortality rates similar to, or lower, than White patients.30 In this study, hospitals attending to a population with more insured patients were more effective at meeting time-to-operating room benchmarks regardless of hospital racial composition, which was consistent with the Kaiser Permanente system where all patients are insured.The results of this study are consistent with prior studies evaluating disparities in orthopedic care on a systems or institutional scale as well as those evaluating disparities on a patient-level.29 It may be that it is more difficult for clinicians to detect disparities that occur as a result of hospital-level systemic issues as opposed to patient-by-patient disparities.The results of this study can explain, at least in part, the results of a survey of the American Association of Orthopaedic Surgery, which found that only 9% of orthopedic surgeons believe that race or ethnicity influences orthopedic care that is provided.27 suggest that racial disparities may be occurring on a health systems level as opposed to a patient-by-patient level. Targeting health care systems for quality improvement and public health initiatives is critical to improving health care equity.The disparities identified in this, and prior investigations, signal an alarming deficiency of the US health care system. While identification of health disparities is a critical first step, understanding reasons for these disparities is critical to progress toward a more equitable health care system. This and other analyses28 that are subject to issues with inconsistent reporting and coding inaccuracies14 or single center studies.31 Furthermore, this analysis included several time-sensitive fractures, both of which demonstrated the same effect in the primary analysis suggesting that this association was more widespread across injuries and not limited to a specific injury type. Additionally, the multicenter nature of this study allowed for evaluation of disparities on a national scale, as opposed to a single center.There are several strengths associated with this study. This study leveraged a high-quality prospectively collected, adjudicated data set that had advantages in validity over prior research using administrative data There are several limitations associated with this study. The results of this study may not be generalizable to all institutions since the data for this study was derived from a prospective study at private nonprofit Level I and II trauma centers with funded research infrastructure. It likely that institutions with a strong research infrastructure are better resourced than institutions without a strong resource infrastructure, suggesting that these results may be even more profound across a broader range of institutions. We used a single model approach for both hip and femur fractures because we were adjusting for the same set of independent variables. This has the benefit of a larger sample size and thus more statistical power, which helps alleviate issues associated with imbalance in the outcome. Furthermore, fracture location (femur vs hip) was included in the multivariate model and there was no association between fracture location and time-to-operating room. Because the PREPARE trial was not designed to assess racial disparities for time-sensitive procedures, the groups were not evenly distributed and there were differences between patient cohorts who met the 24-hour time-to-surgery benchmark and those that did not. However, multivariate regression was used to control for these differences, as is indicated for observational research.This cohort study demonstrated that patients seeking care from an institution with a patient population that was more racially and ethnically diverse and greater proportions of uninsured patients were more likely to experience delays greater than the 24-hour time-to-surgery benchmark, regardless of the individual patient race or ethnicity. Furthermore, institutions that treated a less diverse patient population appeared to be more resilient to the mix of insurance status in their patient population and were able to meet time-to-operating room benchmarks regardless of patient insurance status or population-based insurance mix. While increased delays in association with a poorer payor mix would not be surprising, the fact that delays are only seen in association with a more racially diverse patient population can only be explained by structural health systems issues.


Small bowel polyps exhibit variations related to color, shape, morphology, texture, and size, as well as to the presence of artifacts, irregular polyp borders, and the low illumination condition inside the gastrointestinal GI tract. Recently, researchers developed many highly accurate polyp detection models based on one-stage or two-stage object detector algorithms for wireless capsule endoscopy (WCE) and colonoscopy images. However, their implementation requires a high computational power and memory resources, thus sacrificing speed for an improvement in precision. Although the single-shot multibox detector (SSD) proves its effectiveness in many medical imaging applications, its weak detection ability for small polyp regions persists due to the lack of information complementary between features of low- and high-level layers. The aim is to consecutively reuse feature maps between layers of the original SSD network. In this paper, we propose an innovative SSD model based on a redesigned version of a dense convolutional network (DenseNet) which emphasizes multiscale pyramidal feature maps interdependence called DC-SSDNet (densely connected single-shot multibox detector). The original backbone network VGG-16 of the SSD is replaced with a modified version of DenseNet. The DenseNet-46 front stem is improved to extract highly typical characteristics and contextual information, which improves the model’s feature extraction ability. The DC-SSDNet architecture compresses unnecessary convolution layers of each dense block to reduce the CNN model complexity. Experimental results showed a remarkable improvement in the proposed DC-SSDNet to detect small polyp regions achieving an mAP of 93.96%, F1-score of 90.7%, and requiring less computational time. Recently, small bowel tumors have been the third leading cause of death in the word. Adenomatous polyps formed by glandular tissue are considered as one of the most common cases of colorectal cancer. Contrary to hyperplastic polyps, which have no malignant potential, adenomas are considered precancerous and can transform into malignant structures despite being benign tumors. The prevalence of this disease is expected to rise in the coming years . For thaThe application of a modified version of DenseNet called DenseNet-46 as a backbone and smoothly adapted to the SSD detector to improve its ability for small polyp detection.Based on the inception v4 stem part, the backbone DenseNet-46 front stem is improved, allowing the extraction of highly relevant features and contextual information.To capture enough patterns and representative information, we increased the filter numbers of the first convolution layers in the stem part of the DenseNet-46 backbone from 32 to 64.We omitted the unnecessary convolution layers of each dense block of the DenseNet-46 backbone to reduce the DC-SSDNet model’s complexity and to achieve a faster speed while preserving a lesser computation time.DC-SSDNet adds a couple of new dense and transition blocks to match the structure of SSD that detects targets in images using a single deep neural network.DC-SSDNet introduces additional convolution layers to the multiscale feature pyramid, which is consistent with the traditional SSD.The proposed model is trained from scratch.We conducted several experiments on three well-known datasets in the field  to verify the DC-SSDNet model’s effectiveness for a fair comparison with previously published methods of the literature.This manuscript provides a thorough examination of the benefits and drawbacks of the proposed framework.The following are the primary contributions of this study:The rest of this paper is structured as follows: WCE images exhibit great variations in terms of size, shape, and morphology making the automated polyp detection process more difficult. Handcrafted features have been widely investigated, showing a significant progress for gastrointestinal classification tasks. A previous study  presenteenseNets ). RegardenseNets  presenteenseNets  proposedenseNets  presenteenseNets  presenteenseNets .A deep CNN uses many repeated convolution layers resulting in the bottom-level features being destroyed. Many medical imaging applications utilize DenseNet to improve precision caused by the vanishing gradient in high-level neural networks. DenseNet reuses some not-useful redundant information to concatenate the high-level features with the residual sparse low-level ones, even if it does not propagate them effectively. In this paper, we propose a compact DenseNet-46 as a backbone network of DC-SSDNet by applying certain modifications to DenseNet. The motivation behind the DenseNet-46 stem block design originates from the considerable success of the Inception-v4 structure  in smallThe dense block is the basic unit of the compact DenseNet-46 structure, as shown in After several dense connections, the number of feature maps of the DenseNet-46 grows dramatically. Consistent with the original DenseNet, we designed transition layer blocks to reduce the dimension of previous dense blocks’ features, as depicted in Several feature layers  are added to the original SSD via the base network’s end (VGG16) for object classification and location regression from multiple feature maps at different depths. Inspired by the pyramidal feature hierarchy, the proposed approach adopts the SSD network structure . HoweverThe WCE dataset is from PillCam©COLON 2 polyps . It consIn this study, we performed the training and testing phases using the Colab Pro Plus solution provided by Google, with a maximum RAM of 52 Gb and a disk of 166.83 Gb. The CUDA 8.0.61-1, CuDNN6.0, Keras 2.1.0, Python 3.7, protobuf 3.20.*, h5py 2.10.0, NumPy 1.16.3, TensorFlow 1.14, TensorFlow-GPU 1.14, OpenCV-python, scikit-learn, scikit-image, tqdm, beautifulsoup4, lxml, html5lib, bs4, ipykernel, and OpenCV 3.1 packages were used to implement the algorithm. For all used datasets, an aspect ratio between one and two was adopted depending on the small polyp regions’ true bounding boxes.Many training hyperparameters were adjusted to strengthen the DC-SSDNet capability for detecting polyp abnormalities in WCE and colonoscopy frames as shown in −8). The learning rate was initially set to 0.0001, and the learning rate decay policy differed slightly from the original SSD with a drop of 0.5 and an epoch drop of 10, which allowed the network to converge by controlling its learning rate. If the epoch number was less than 80, the learning rate was 0.0001, 0.00001 if the epoch number was less than 100, and 0.000001 otherwise. Recent state-of-the-art approaches targeting polyp detection were investigated in this study to compare them with the proposed model [We trained the proposed model from scratch to not rely on any model pretrained on classification tasks to initialize the network, as commonly known visual purposes of classification and detection are distinct. For training DC-SSDNet, we used adaptive moment estimation (Adam) as the optimizer  and frames per second (FPS), which are well-known in target object detection. The mAP represents the average of all object categories’ average precision (AP), expressed as:The precision indicator represents a measure of exactness and the recall a measure of completeness. The expected overlap between the ground-truth bounding box annotated by the experts and the predicted one produced by the network expresses the intersection over union (IoU). The following equations formulate the precision, recall, and F1-score indicators as follows: work of .On the WCE, CVC-ClinicDB, and ETIS-Larib colonoscopy datasets, we performed an ablation study to investigate the impact of each component of the DC-SSDNet detector on performance. We trained the detector-based SSDs on the WCE, CVC-ClinicDB, and ETIS-Larib combined training set and tested on each dataset separately as depicted from We compared the performance of the proposed DC-SSDNet approach to the most prominent networks in the literature-based SSDs models, YOLOv3 and Faster R-CNN, as shown in The main objective of this study was to show the impact of the condensed model in highlighting small polyp regions for the detection and localization tasks on the WCE, Etis-Larib, and CVC-ClinicDB test sets. Polyp detection aims at selecting polyp areas and ignoring normal parts, feces, artifacts, and water jet sprays to clean the colon. Some examples of polyp detection of the SSD model and DC-SSDNet network on the employed datasets are depicted in This work suggested an improved SSD detector based on a redesigned DenseNet (DC-SSDNet) emphasizing polyp detection in the WCE, Etis-Larib, and CVC-ClinicDB datasets. Many researchers in the field use their own datasets due to medical imaging ethics and the lack of publicly available and annotated WCE polyp datasets. Thus, their results may suffer from subjectivity. The DC-SSDNet network aimed to exploit the potential of feature reuse as opposed to relearning features in later layers. The proposed approach used stacked dense and transition blocks instead of simple convolution layers with a capacity of the SSD network to be more compact while going deeper, yielding condensed models that were easy to train and highly parameter-efficient. The compact DenseNet-46 compressed unnecessary convolution layers of each dense block to reduce the amount of feature redundancy, resulting in fewer overall parameters and faster training times. The small polyp areas’ visual appearance was modeled by directly connecting layers throughout the network and generating novel pyramidal feature maps. While training data from scratch, the DC-SSDNet detector achieved comparable, if not superior, performance in mAP compared to other state-of-the-art pretrained models with real-time processing speed on the WCE and public datasets (CVC-ClinicDB and Etis-Larib). Artifacts and other factors may degrade performance and negatively affect the detection process. We will conduct further studies in the future around this shortcoming while maintaining the smaller running time. Rather than proposing polyp identification tasks on WCE and colonoscopy frames, future studies will investigate hybrid architectures to present a unique detection method for video colonoscopy. Furthermore, it would be interesting to use a hybrid 2D/3D architecture and assess its performance by employing other promising backbones.


During the current pandemic, the vast majority of COVID-19 patients experienced mild symptoms, but some had a potentially fatal aberrant hyperinflammatory immune reaction characterized by high levels of IL-6 and other cytokines. Modulation of this immune reaction has proven to be the only method of reducing mortality in severe and critical COVID-19. The anti-inflammatory drug baricitinib (Olumiant) has recently been strongly recommended by the WHO for use in COVID-19 patients because it reduces the risk of progressive disease and death. It is a Janus Kinase (JAK) 1/2 inhibitor approved for rheumatoid arthritis which was suggested in early 2020 as a treatment for COVID-19. In this review the AI-assisted identification of baricitinib, its antiviral and anti-inflammatory properties, and efficacy in clinical trials are discussed and compared with those of other immune modulators including glucocorticoids, IL-6 and IL-1 receptor blockers and other JAK inhibitors. Baricitinib inhibits both virus infection and cytokine signalling and is not only important for COVID-19 management but is “non-immunological”, and so should remain effective if new SARS-CoV-2 variants escape immune control. The repurposing of baricitinib is an example of how advanced artificial intelligence (AI) can quickly identify new drug candidates that have clinical benefit in previously unsuspected therapeutic areas.  The COVID-19 pandemic has caused the death of approximately 6 million people, with a case fatality rate which may be as high as 20% in those over 80 years old . A . A 56]. By focusing on anti-inflammatory and viral infection mechanisms as described in , the knoThe combination of the oncology therapeutics sunitinib and erlotinib was previously shown to reduce the infectivity of a wide range of viruses, including Hepatitis C virus, Dengue virus, Ebola virus, and respiratory syncytial virus ,58. HoweAfter publishing the output of this AI-augmented research, the mechanistic predictions were validated, confirming that baricitinib inhibited signalling by a range of cytokines associated with the COVID-19 hyperinflammation. The anti-inflammatory effects of JAK inhibitors in general are summarised in ref. , and theBaricitinib has other advantages including an oral once per day formulation, a predominantly renal route of clearance and low plasma protein binding. These properties suggested that baricitinib could be readily dosed with the antivirals being tested at the start of the pandemic, since they were largely cleared through liver metabolism. This enabled the testing of a combination with remdesivir in the ACTT-1 trial as well as with drugs being used as standard care.The first clinical test of baricitinib in COVID-19 was in four patients in Milan, all of whom recovered well . ImportaThese data helped to convince the National Institute for Allergy and Infectious Diseases of the NIH that baricitinib should be tested in a randomised trial. Since remdesivir had already shown a positive outcome in the ACTT-1 trial, , bariciThe outcome of the ACTT-2 trial with over 1000 patients was a significant reduction in mortality and accelerated recovery in patients treated with baricitinib plus remdesivir compared with those treated with remdesivir alone. These effects were seen most strongly in those requiring supplemental oxygen  or non-invasive ventilation  at baseline. Both groups showed a 40% reduction in mortality . Based oStudies of new therapies in at-risk individuals with early stage COVID-19  have been difficult to execute because of the need for strict patient isolation, which precludes careful monitoring. Advanced contactless monitoring methods are being developed to enable such trials to be undertaken , includiIt is probable that this approach of targeting the host pathways subverted by the virus could be applied to other infections. For instance, flavivirus  infections also show an early relatively mild disease which, in some cases, is followed by an aberrant hyperinflammation with vasculopathy. The Dengue viruses (and their NS1 proteins) access cells predominantly through CME , so antiLong-term treatment (months to years) with JAK inhibitors is associated with significant side effects in a small number of patients . These iThe WHO has a live document published in the BMJ in which its recommended therapies for COVID-19 are summarised . In summDuring this period, two other JAK inhibitors were tested in randomised COVID-19 trials, ruxolitinib and tofacitinib, neither of which was predicted to inhibit the NAK enzymes at therapeutic plasma concentrations. Early observational trials of the JAK1/3 inhibitor tofacitinib ,82 and tLow-dose dexamethasone has been widely used since the interim analysis of the RECOVERY clinical trial revealed its beneficial effects . Later aDuring the hyperinflammation phase of COVID-19, IL-6 is frequently the most highly elevated cytokine in the plasma. This led to the testing of the anti-IL6 receptor monoclonal antibody tocilizumab in multiple trials in 2020, leading to its approval for the treatment of COVID-19 in the US, UK, and EU, and being recommended by the WHO. Significant improvements in mortality were seen in the REMAP-CAP  and RECOIL-1β is another cytokine frequently elevated in COVID-19 patients, and the anti-IL-1 receptor antibody anakinra has been tested in clinical trials. In a meta-analysis covering 895 patients in mostly observational studies, a significant effect on mortality was seen when dosed in the absence of steroids HR 0.32). Surprisingly, this analysis suggested anakinra was of no further benefit when given with steroids . In a hi. SurprisDespite the variability in the outcomes of the studies in In summary, baricitinib has shown the greatest mortality benefit in large numbers of COVID-19 patients in randomised clinical trials  while toFrom all of this, it is clear that, as recommended by the WHO, baricitinib or tocilizumab with low-dose glucocorticoids are the most effective therapies currently available for severely ill patients. It is also probable that the success seen with the combined antiviral and anti-inflammatory effects of baricitinib could lead to the effective treatment of other viral diseases  characterised by clathrin-mediated infection of target cells and hyperinflammation.In conclusion, baricitinib has proven to be the best available drug for COVID-19, limiting the inflammatory phase of the disease and reducing the risk of progression to advanced disease. It has been approved by the World Health Organization and has entered standard of care guidelines in many centres. It represents an example of the power of AI to identify useful drugs and is a model for the rapid discovery of potentially useful drugs in future pandemics. Finally, baricitinib is an example of a drug that limits disease independent of host immunity, and such drugs will be essential if new, highly pathogenic SARS-CoV-2 variants evade existing immunity or if similar viral pandemics occur in the future.


Indocyanine green (ICG) fluorescence angiography (FA) is used for several purposes in general surgery, but its use in bariatric surgery is still debated. The objective of the present pilot study is to evaluate the intraoperative utility of ICG-FA during bariatric surgery in order to focus future research on a reliable tool to reduce the postoperative leak rate.2  underwent bariatric surgery with ICG-FA in our center. Three mL of ICG diluted with 10 cc sterile water were intravenously injected after gastric tube creation during laparoscopic sleeve gastrectomy (LSG) and after the gastric pouch and gastro-jejunal anastomosis creation during laparoscopic gastric by-pass (LGB). For the ICG-FA, Karl Storz Image 1S D-Light system  placed at a fixed distance of 5 cm from the structures of interest and zoomed vision modality were used to identify the vascular supply. The perfusion pattern was assessed by the surgical team according to a score. The score ranged from 1 (poor vascularization) to 5 (excellent vascularization) based on the intensity and timing of fluorescence of the vascularized structures.Thirteen patients  with median age of 52 years  and preoperative median body mass index of 42.6 kg/mFom January 2021 to February 2022, six patients underwent LSG (46.2%), three patients underwent LGB (23.1%), and four patients underwent re-do LGB after LSG (30.8%). No adverse effects to ICG were observed. In 11 patients (84.6%) ICG-FA score was 5. During two laparoscopic re-do LGB, the vascular supply was not satisfactory (score 2/5) and the surgical strategy was changed based on ICG-FA (15.4%). At a median follow-up of five months postoperatively, leaks did not occur in any case.ICG-FA during bariatric surgery is a safe, feasible and promising procedure. It could help to reduce the ischemic leak rate, even if standardization of the procedure and objective fluorescence quantification are still missing. Further prospective studies with a larger sample of patients are required to draw definitive conclusions. Bariatric surgery has strong evidence of efficacy and safety in the management of obesity and related comorbidities , 2. LapaStaple line dehiscence and leak after LSG and of the gastric pouch or gastro-jejunal anastomosis after LGB are the most severe and feared complications after bariatric surgery , 7, withSeveral techniques have been described to prevent and to decrease the ischemic and mechanical leak incidence, including manual oversewing of the staple line or the use of sealants, or the use of staple line buttressing material  , 17. HowSome authors have recently used the intravenous administration of indocyanine green (ICG) to evaluate the real-time tissue perfusion during LSG, without achieving definitive results –24. ICG The objective of the present pilot study is to evaluate the intraoperative utility of ICG-FA during bariatric surgery in order to focus future research on a reliable tool to reduce the postoperative leak rate.This is a prospective observational pilot study. Institutional review board approval (RP120172B794FE16) and informed consent from all participants were obtained.Fom January 2021 to February 2022, thirteen consecutive patients underwent bariatric surgery with ICG-FA in our center .All patients were candidate for bariatric surgery according to the Italian Society for Obesity and Metabolic Surgery (SICOB) guidelines as well as the European Association for Endoscopic Surgery (EAES) guidelines, as previously reported , 34.Preoperative endoscopy with biopsies for Helicobacter Pylori (HP) detection was performed routinely –36. In cThe presence of signs of gastroesophageal reflux disease (GERD) was evaluated by endoscopy and patients’ symptoms were investigated by the Modified Italian Gastroesophageal reflux disease - Health-Related Quality of Life (MI-GERD-HRQL) questionnaire in all patients . When prRoutine preoperative manometry, pH-metry and Rx-esophagogram were performed in accordance with Severe Acute Respiratory Syndrome CoronaVirus 2 (SARS-CoV-2) restrictions and patient compliance , 42. A DBased on this preoperative workup, after multidisciplinary evaluation and after obtaining the patients’ informed consent, patients without GERD underwent LSG and patients with GERD underwent LGB. Patients with GERD after LSG, with or without weight regain, underwent RLGB.In any case, if a diagnosis of HH was made intraoperatively, this was repaired with hiatoplasty according to guidelines , with meSurgery was performed by the same surgeon (A.M.P.) and surgical team.LSG was performed with a standardized technique as previously reported –36. DoubAfter gastric tube creation during LSG and after gastric pouch creation and gastro-jejunal anastomosis during LGB and RLGB, three mL of ICG  diluted with 10 cc of sterile water were injected intravenously. For the ICG-FA, Karl Storz Image 1S D-Light system  placed at a fixed distance of five centimeters from the structures of interest was used to identify the vascular supply.During LSG, the rationale to ICG administration after gastric tube creation is related to the fact that to obtain sufficient weight loss, the stomach must be divided on the guide of the bougie. So ICG-FA is useful to decide the site of a manual stitch on the section line, if ischemic.The perfusion pattern was evaluated by the surgical team in real time, immediately after ICG injection and was assessed according to a score ranging from 1 (poor vascularization) to 5 (excellent vascularization) based on the intensity and rapidity of fluorescence appearance in the observed structures .A score of 1–2 was considered inadequate and led to a change in surgical strategy, while a score of 3–5 was considered satisfactory, and the surgical strategy did not change. A methylene blue test is routinely performed after ICG-FA during LGB and RLGB. A routine Rx-esophagogram is performed on the third postoperative day after LGB and RLGB.Preoperative variables , intraoperative variables  and postoperative variables  while categorical variables are expressed as frequencies and percentages.Six (46.2%), three (23.1%) and four (30.8%) patients underwent LSG, LGB and RLGB, respectively. Ten patients underwent associated procedures (76.9%): hiatoplasty in 5 cases (38.5%), cholecystectomy in 3 (23.1%), lysis of adhesions from previous abdominal surgery in 1 (7.7%) and liver biopsy for severe hepatic steatosis in 1 (7.7%). All patients underwent preoperative multidisciplinary surgical-hepatological-radiological evaluation.ICG-FA was performed in all patients with no adverse events, and without significant increase in operative time. In 11 patients ICG-FA score was 5/5 (84.6%). During two RLGB procedures, ICG-FA score was 2/5, and the surgical strategy was changed (15.4%).2), the ICG-FA demonstrated poor blood supply of a small area of the gastro-jejunal anastomosis (score 2/5), so reinforcement stitches were placed along the staple line.In a 48-year-old woman with symptomatic GERD after LSG and weight regain (preoperative BMI 42.6 kg/m2) after LSG the ICG-FA showed inadequate perfusion at the periphery of the gastric pouch (score 2/5). The gastric pouch was thus reshaped with the stapler, and the subsequent ICG-FA confirmed good perfusion of the residual gastric pouch (score 5/5).In a 64-year-old woman with mild GERD and weight regain : two patients with fever, and one with wound infection, all treated by antibiotic therapy (Clavien-Dindo grade II).No leaks were observed at postoperative Rx -esophagogram. Median hospital stay was four days  and mortality was nil. At a median follow-up of five months  leaks did not occur.The aim of this pilot study was to investigate if the intraoperative use of ICG-FA during bariatric surgery is a useful tool to reduce the postoperative ischemic leak rate. For this purpose, we prospectively analyzed the preliminary results of thirteen patients undergoing LSG or LGB with ICG-FA in our center. Even if leaks were not observed in any case, the use of ICG-FA changed the surgical strategy intraoperatively in two out of thirteen patients (15.4%), who underwent re-do surgery.Leak of the staple line and of the gastric pouch or the gastro-jejunal anastomosis after LSG or LGB, respectively, are the most severe complications and the second most common cause of death after bariatric surgery , 7, 47. According to the mechanical theory, the intraluminal pressure due to pyloric conservation after LSG exceeds the strength of the staple-line, resulting in a leak . MechaniOn the other hand, according to the ischemic theory, leaks are due to localized ischemia which occurs most frequently in the “critical area” at the level of the angle of His after gastroepiploic and short gastric vessels ligation . IschemiSeveral modalities of intraoperative leak diagnosis  have been described, with discordant results , 53, 60.Some authors have recently used the intraluminal or intravenous administration of ICG to evaluate real-time tissue perfusion and to assess the integrity of the staple line during bariatric surgery, without, however, achieving definitive results yet , 27, 61.In the literature, the use of ICG-FA during bariatric surgery is reported only in three articles –24, and Frattini et al. used ICG-FA in 15 patients undergoing LSG . They reDi Furia et al. performed LSG with ICG-FA in 45 patients with the aim of clarifying the exact pathogenesis of gastric leak supporting the ischemic theory and to evaluate if ICG-FA could adequately estimate the ischemic area along the staple line and to prevent leaks . They deIn our opinion, although this conclusion can be applied to LSG, we do not believe it is entirely true for revisional bariatric surgery. To the best of our knowledge, this is the first study evaluating the role of ICG-FA during primary or revisional LGB. In this series the only two cases in which the ICG-FA determined a change in surgical strategy were RLGB, suggesting how the ICG could have a higher detection rate of ischemic areas during revisional surgery.With the same goal, Ortega et al. performed ICG-FA before the gastric division during LSG to identify the variable blood supply patterns of the gastroesophageal junction and the ICG-FA was repeated at the end of the procedure to ensure that all the pertinent blood vessels were preserved . They haFinally, another use of ICG described in the literature is the intraluminal gastric injection . Kalmar No adverse effects of ICG administration were observed in the reported series as in the present analysis, which therefore is confirmed to be a safe dye , 24, 27.In the literature a wide heterogeneity regarding the doses and timing of ICG administration is reported , as wellIn the present study the use of ICG-FA allowed to identify an ischemic area leading to a change in the surgical strategy in half of the patients undergoing revisional bariatric surgery. Although this result could be influenced by the small sample size, this is not yet described in the literature, so if our preliminary results will be confirmed, the use of ICG-FA could represent a decisive weapon in the prevention of ischemic leak in revisional bariatric surgery.In conclusion, based on the present study, ICG-FA during bariatric surgery is a safe and feasible procedure. Its use could reduce the ischemic leak rate, especially in patients undergoing revisional bariatric surgery even if standardization of the procedure and objective fluorescence quantification are still missing. Definitive data from this pilot study increasing the number of patients included, and further prospective studies with a larger number of patients are required to draw definitive conclusions.