The number of proposed prognostic models for coronavirus disease 2019 (COVID-19) is growing rapidly, but it is unknown whether any are suitable for widespread clinical implementation.We independently externally validated the performance of candidate prognostic models, identified through a living systematic review, among consecutive adults admitted to hospital with a final diagnosis of COVID-19. We reconstructed candidate models as per original descriptions and evaluated performance for their original intended outcomes using predictors measured at the time of admission. We assessed discrimination, calibration and net benefit, compared to the default strategies of treating all and no patients, and against the most discriminating predictors in univariable analyses.We tested 22 candidate prognostic models among 411 participants with COVID-19, of whom 180 (43.8%) and 115 (28.0%) met the endpoints of clinical deterioration and mortality, respectively. Highest areas under receiver operating characteristic (AUROC) curves were achieved by the NEWS2 score for prediction of deterioration over 24 h , and a novel model for prediction of deterioration <14 days from admission . The most discriminating univariable predictors were admission oxygen saturation on room air for in-hospital deterioration , and age for in-hospital mortality . No prognostic model demonstrated consistently higher net benefit than these univariable predictors, across a range of threshold probabilities.Admission oxygen saturation on room air and patient age are strong predictors of deterioration and mortality among hospitalised adults with COVID-19, respectively. None of the prognostic models evaluated here offered incremental value for patient stratification to these univariable predictors. Oxygen saturation on room air and patient age are strong predictors of deterioration and mortality, respectively, among hospitalised adults with COVID-19. None of the 22 prognostic models evaluated in this study adds incremental value to these univariable predictors.https://bit.ly/2Hg24TO Coronavirus disease 2019 (COVID-19), caused by severe acute respiratory syndrome coronavirus-2 (SARS-CoV-2), causes a spectrum of disease ranging from asymptomatic infection to critical illness. Among people admitted to hospital, COVID-19 has a reported mortality of 21%–33%, with 14%–17% requiring admission to high dependency or intensive care units (ICUs) . Rapid aAs a result, there has been global interest in developing prediction models for COVID-19 . These iWe used a published living systematic review to identify all candidate prognostic models for COVID-19 indexed in PubMed, Embase, Arxiv, medRxiv or bioRxiv until May 5, 2020, regardless of underlying study quality . We inclOur study is reported in accordance with transparent reporting of a multivariable prediction model for individual prognosis or diagnosis (TRIPOD) guidance for external validation studies . We inclData were collected by direct extraction from electronic health records, complemented by manual curation. Variables of interest in the dataset included demographics , comorbidities , clinical observations, laboratory measurements, radiology reports and clinical outcomes. Each chest radiograph was reported by a single radiologist, who was provided with a short summary of the indication for the investigation at the time of request, reflecting routine clinical conditions. Chest radiographs were classified using British Society of Thoracic Imaging criteria, and using a modified version of the Radiographic Assessment of Lung Edema (RALE) score , 15. ForFor models that used ICU admission or death, or progression to “severe” COVID-19 or death, as composite endpoints, we used a composite “clinical deterioration” endpoint as the primary outcome. We defined clinical deterioration as the initiation of ventilatory support  or death, equivalent to World Health Organization (WHO) Clinical Progression Scale ≥6 . This deParticipants were followed up clinically to the point of discharge from hospital. We extended follow-up beyond discharge by cross-checking National Health Service (NHS) Spine records to identify reported deaths post-discharge, thus ensuring >30 days’ follow-up for all participants.For each prognostic model included in the analyses, we reconstructed the model according to the authors’ original descriptions, and sought to evaluate the model discrimination and calibration performance against our approximation of their original intended endpoint. For models that provide online risk calculator tools, we validated our reconstructed models against original authors’ models by cross-checking our predictions against those generated by the web-based tools for a random subset of participants.versus observed risk using LOESS-smoothed plots, and by quantifying calibration slopes and calibration-in-the-large (CITL). A perfect calibration slope should be 1; slopes <1 indicate that risk estimates are too extreme, while slopes >1 reflect risk estimates not being extreme enough. Ideal CITL is 0; CITL>0 indicates that predictions are systematically too low, while CITL<0 indicates that predictions are too high. For models with points-based scores, we assessed calibration visually by plotting model scores versus actual outcome proportions. For models that provide probability estimates, but where the model intercept was not available, we calibrated the model to our dataset by calculating the intercept when using the model linear predictor as an offset term, leading to perfect CITL. This approach, by definition, overestimated calibration with respect to CITL, but allowed us to examine the calibration slope in our dataset.For all models, we assessed discrimination by quantifying the area under the receiver operating characteristic curve (AUROC) . For modWe also assessed the discrimination of each candidate model for standardised outcomes of 1) our composite endpoint of clinical deterioration and 2) mortality across a range of pre-specified time horizons from admission  by calculating time-dependent AUROCs (with cumulative sensitivity and dynamic specificity) . The rata priori, based on clinical knowledge and existing data, for predicting our composite endpoints of clinical deterioration and mortality . The a priori predictors of interest examined in this analysis were age, clinical frailty scale, oxygen saturation at presentation on room air, C-reactive protein level and absolute lymphocyte count [To further benchmark the performance of candidate prognostic models, we then computed AUROCs for a limited number of univariable predictors considered to be of highest importance te count , 19.rmda package in R [Decision curve analysis allows the clinical utility of candidate models to be assessed, and is dependent on both model discrimination and calibration . We perfage in R .mice package in R [We handled missing data using multiple imputation by chained equations , using tage in R . All varage in R . A totalage in R .All analyses were conducted in R .a priori univariable predictors using restricted cubic splines, with three knots. Finally, we estimated optimism for discrimination and calibration parameters for the a priori univariable predictors using bootstrapping (1000 iterations), using the rms package in R [We recalculated discrimination and calibration parameters for each candidate model using 1) a complete case analysis (in view of the large amount of missingness for some models), 2) excluding patients without PCR-confirmed SARS-CoV-2 infection and 3) excluding patients who met the clinical deterioration outcome within 4 h of arrival at hospital. We also examined for non-linearity in the age in R .The pre-specified study protocol was approved by East Midlands - Nottingham 2 Research Ethics Committee .supplementary figure S1 and supplementary table S1.We identified 37 studies describing prognostic models, of which 19 studies (including 22 unique models) were eligible for inclusion (supplementary figure S2). The median age of the cohort was 66 years (interquartile range (IQR) 53–79 years), and the majority were male . supplementary figure S3). Most of the variables required for calculating the 22 prognostic model scores were available among the vast majority of participants. However, admission lactate dehydrogenase was only available for 183 out of 411 (44.5%) and D-dimer was only measured for 153 out of 411 (37.2%), resulting in significant missingness for models requiring these variables (supplementary figure S4).During the study period, 521 adults were admitted with a final diagnosis of COVID-19, of whom 411 met the eligibility criteria for inclusion . These analyses showed that AUROCs generally declined with increasing time horizons. Admission peripheral oxygen saturation on room air was the strongest predictor of in-hospital deterioration , while age was the strongest predictor of in-hospital mortality .Next, we sought to compare the discrimination of these models for both clinical deterioration and mortality across the range of time horizons, benchmarked against preselected univariable predictors associated with adverse outcomes in COVID-19 , 19. We i.e. oxygen saturation on air) or mortality (i.e. patient age) to stratify treatment (supplementary figure S8). Although all prognostic models showed greater net benefit than treating all patients at the higher range of threshold probabilities, none of these models demonstrated consistently greater net benefit than the most discriminating univariable predictor, across the range of threshold probabilities  to the strategies of treating all patients, treating no patients and using the most discriminating univariable predictor for either deterioration (bilities .supplementary table S2). Visual examination of associations between the most discriminating univariable predictors and log odds of deterioration or death using restricted cubic splines showed no evidence of non-linear associations (supplementary figure S9). Finally, internal validation using bootstrapping showed near zero optimism for discrimination and calibration parameters for the univariable models (supplementary table S3).Recalculation of model discrimination and calibration metrics for predicting the original intended endpoint using 1) a complete case analysis, 2) excluding patients without PCR-confirmed SARS-CoV-2 infection and 3) excluding patients who met the clinical deterioration outcome within 4 h of arrival to hospital revealed similar results to the primary multiple imputation approach, though discrimination was noted to be lower overall when excluding early events (In this observational cohort study of consecutive adults hospitalised with COVID-19, we systematically evaluated the performance of 22 prognostic models for COVID-19. These included models developed specifically for COVID-19, along with existing scores in routine clinical use prior to the pandemic. For prediction of clinical deterioration or mortality, AUROCs ranged from 0.56–0.78. NEWS2 performed reasonably well for prediction of deterioration over a 24-h interval, achieving an AUROC of 0.78, while the Carr “final” model  also hadWhile previous studies have largely focused on novel model discovery, or evaluating a limited number of existing models, this is the first study to our knowledge to evaluate systematically identified candidate prognostic models for COVID-19. We used a comprehensive living systematic review  to identA weakness of the current study is that it is based on retrospective data from a single centre, and therefore cannot assess between-setting heterogeneity in model performance. Second, owing to the limitations of routinely collected data, predictor variables were available for varying numbers of participants for each model, with a large proportion of missingness for models requiring lactate dehydrogenase and D-dimer measurements. We therefore performed multiple imputation, in keeping with recommendations for development and validation of multivariable prediction models, in our primary analyses . FindingDespite the vast global interest in the pursuit of prognostic models for COVID-19, our findings show that none of the COVID-19-specific models evaluated in this study can currently be recommended for routine clinical use. In addition, while some of the evaluated models that are not specific to COVID-19 are routinely used and may be of value among inpatients , 44, peoWe conclude that baseline oxygen saturation on room air and patient age are strong predictors of deterioration and mortality, respectively. None of the prognostic models evaluated in this study offer incremental value for patient stratification to these univariable predictors when using admission data. Therefore, none of the evaluated prognostic models for COVID-19 can be recommended for routine clinical implementation. Future studies seeking to develop prognostic models for COVID-19 should consider integrating multi-centre data to increase generalisability of findings, and should ensure benchmarking against existing models and simpler univariable predictors.10.1183/13993003.03498-2020.Supp1Please note: supplementary material is not edited by the Editorial Office, and is uploaded as it has been supplied by the author.ERJ-03498-2020.SupplementSupplementary material 10.1183/13993003.03498-2020.Shareable1This one-page PDF can be shared freely online.ERJ-03498-2020.ShareableShareable PDF


The activity of the entire population would then generate the appropriate synergy, depending on the location of sensory stimulation. Mathematical modeling has demonstrated that this hypothesis is feasible. Here, we provide one test of this hypothesis by injecting two fluorescent retrograde tracers into the regions of knee-extensor motoneurons  and hip-extensor motoneurons . We found that there were double-labeled interneurons, which projected their axons to both locations. The dual-projecting interneurons were widely distributed rostrocaudally, dorsoventrally, and mediolaterally within the hindlimb enlargement and pre-enlargement spinal segments examined. The existence of such dual-projecting interneurons is consistent with the hypothesis that they contribute to generating the knee-hip synergy for pocket scratching. The dual-projecting interneurons, however, were only about 1% of the total interneurons projecting to each location, which suggests that they might be one of several contributors to the appropriate knee-hip synergy. Indirect projections to both motor pools and/or knee extensor-dedicated interneurons might also contribute. There is evidence for dual-projecting spinal interneurons in frogs and mice as well, suggesting that they may contribute to limb motor control in a variety of vertebrates.The central nervous system (CNS) may simplify control of limb movements by activating certain combinations of muscles together, i.e., muscle synergies. Little is known, however, about the spinal cord interneurons that activate muscle synergies by exciting sets of motoneurons for different muscles. The turtle spinal cord, even without brain inputs and movement-related sensory feedback, can generate the patterns of motoneuron activity underlying forward swimming, three forms of scratching, and limb withdrawal. Spinal interneurons activated during scratching are typically activated during all three forms of scratching, to different degrees, even though each form of scratching has its own knee-hip synergy. Such spinal interneurons are also typically activated rhythmically during scratching motor patterns, with hip-related timing. We proposed a hypothesis that such interneurons that are most active during rostral scratch stimulation project their axons to both knee-extensor and hip- The central nervous system (CNS) may simplify limb movement coordination through use of muscle synergies, i.e., simultaneous activation of certain combinations of motoneurons and thereby muscles  8–10 and sacral (S) 1–2. The somata of knee-extensor motoneurons and hip-flexor motoneurons largely overlap rostrocaudally, but the somata of knee-extensor and hip-extensor motoneurons do not , were anesthetized and surgically prepared as described previously . The micropipette was positioned 70% of the way laterally from the midline to Lissauer’s tract and lowered to a depth of 1,200 μm for the rD9 injection and 1,000 μm for the S1 injection to target motoneurons in the ventral horn  using 10–30 psi and 10–30 ms per pulse. Pressure pulses were repeated, with 3-min intervals between pulses, until all of the dye was ejected from the micropipette, ~1.0–1.5 μl. After the injection, the micropipette was left in place for 5 min before withdrawing it. Then, the second location was injected using the same procedure but with the other dye.Micropipettes were pulled from 1.0 mm O.D. borosilicate glass containing a filament , using a P-97 puller (Sutter), and backfilled with either 5% Alexa Fluor 488-dextran amine  or 5% Alexa Fluor 568-dextran amine  Following a survival period of 21–28 days, during which animals were housed in tubs partly filled with water and tilted to facilitate basking, with 12 h-on/12 h-off incandescent and UV lamps, and fed twice weekly with aquatic turtle food pellets , each animal was deeply anesthetized with pentobarbital  and mCherry, Texas Red (Chroma IN007115) filter sets and an Olympus DP70 camera. Single-labeled somata did not visibly fluoresce using the other filter set. Somata were counted as retrogradely labeled neurons if they were outside the cluster of labeled somata at the injection site e.g., , clearlyvia their dendrites. Outside this cluster, labeled neurons were found both ipsilaterally and contralaterally, both rostral and caudal to the injection site, as far as 3 cm away. Even within a small region of the gray matter, some neurons could be found that were labeled with one dye only, some with the other dye only, and still others with both dyes , or the dye .The total number of double-labeled neurons in each animal ranged from 3 to 78, with a mean of 19.4 ± 26.5 and a median of 9. The number of double-labeled neurons was higher in animals with more single-labeled neurons, presumably because more dye had been successfully injected in those animals. Thus, we chose to focus on the percentages of retrogradely labeled neurons that were double-labeled, rather than the absolute numbers. In these six experiments, the mean total percentages of double-labeled neurons were 1.18 ± 0.80% of neurons labeled by the rD9 injection (median: 1.1%) and 0.77 ± 0.39% of neurons labeled by the S1 injection (median: 0.66%).For these six animals, the gray-matter soma locations of all double-labeled neurons were charted on schematic cross-sections for each of the seven spinal cord segments examined: the two pre-enlargement segments, D6 and D7, and the five hindlimb enlargement segments, D8-D10 and S1 . Double-The rostrocaudal distributions of retrogradely labeled neurons appeared to differ for the rD9 and S1 injections . NeuronsIn contrast to the rostrocaudal distributions, the dorsoventral distributions of retrogradely labeled neurons were essentially identical for the rD9 and S1 injections . Most nevia axon terminals.The total number of retrogradely labeled interneurons in each animal in the current study was on average greater than the number we previously achieved injecting horseradish peroxidase (HRP) into the same locations  contains interneuron somata that project their axons to each of these regions, though not necessarily to both  is actually generated by such a circuit. The conclusions of our anatomical experiments are limited by our not knowing: (1) that these interneurons are synaptically connected to the motoneurons we found axonal projections to; and (2) that these interneurons are strongly activated during pocket scratching. Obtaining such data would probably require much more challenging experiments in which interneurons are recorded intracellularly along with two kinds of motoneurons they synapse onto. It is also not clear that the relatively small number of such dual-projecting interneurons we found here  would be sufficient to generate the appropriate knee-hip synergy, even if they are most strongly activated during pocket scratching and synapse onto motoneurons. Thus, if indeed the pocket scratch knee-hip synergy is generated by interneurons with dual outputs to the knee and hip motoneurons, many such interneurons might indirectly affect one or both motor pools , via direct or indirect projections) along with additional interneurons that are dedicated to knee-extensor control, as suggested by the mathematical exploration of the dual-projection circuit (Snyder and Rubin, It is also possible that dual-projecting interneurons that are rhythmically active with hip-related timing contribute to generating the pocket scratch knee-hip synergy (either via dual axon projections (Levine et al., Dual-projecting spinal interneurons could potentially play a role in previously described muscle synergy generation for a variety of limb movements in diverse vertebrates. In adult frogs, extracellular recording of intermediate zone interneurons in the rostral part of the hindlimb enlargement along with spike-triggered averaging of electromyograms has shown that there are spinal interneurons that mono- or disynaptically activate different motor pools, which could contribute to muscle synergy generation (Hart and Giszter, The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation, to any qualified researcher.The animal study was reviewed and approved by the Institutional Animal Care and Use Committee of the University of Oklahoma.AB designed the experiments. KN and TS performed the experiments. TG performed the imaging for figures. KN, TS, and AB analyzed the data. KN and AB wrote the article.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Dear Editor,Coronavirus disease 2019 (COVID-19) is associated with severe systemic inflammation and important elevation of fibrinogen and D-dimers that has been associated with a poor prognosis [We retrospectively reviewed characteristics of patients with confirmed SARS-CoV-2 infection and acute PE who were admitted to our tertiary ICU, which serves as an ECMO referral center for the Greater Paris. In accordance with the ethical standards of French legislation, only non-opposition of patient’s surrogate for utilization of the deidentified data was obtained. The ICU database was registered with the national data protection authority (CNIL 1950673). No analysis for statistical significance was performed given the descriptive nature of the study.From February 25, 2020, to April 6, 2020, 51 patients with confirmed SARS-CoV-2 infection were treated in our ICU, of whom 8 (16%) had confirmed severe PE. Patients’ main characteristics are described in Table We describe a series of 8 critically ill patients with massive PE following COVID-19 infection. Four of these patients developed PE while on VV-ECMO for severe ARDS, a condition that was not reported in the 156 patients included in the EOLIA trial  who receMassive PE in COVID-19 patients may be the consequence of sepsis-induced disseminated intravascular coagulation or to a specific procoagulant state caused by inflammation or virus-induced endothelial dysfunction . ImportaOur observation has potential major clinical implications. First, higher level of anticoagulation might be considered in patients with the most severe forms of the disease, those with high D-dimers and, contrarily to our previous recommendation, in patients supported by VV-ECMO . Second,This case series has several limitations. It is a small single-center case series of critically ill patients, we did not compare clinical and biological characteristics of patients with or without PE, and PE incidence could not be accurately estimated. However, we think that physicians should be warned about the occurrence of severe and potentially fatal PE in COVID-19 patients.Additional file 1:


To evaluate whether routine triple-time-point FDG PET/CT imaging improves the detection of liver metastasis not visualized on initial imaging. To our knowledge, no triple-time-point imaging has been reported. This retrospective study included total 310 patients with various malignancies who underwent PET/CT scans. Triple-time-point imaging including the liver was obtained. The comparison between negative and positive liver lesions on delayed imaging for patients with initial negative imaging were analyzed. Of the 310 patients, 286 did not exhibit liver lesions on initial imaging, but six of the 286 patients exhibited lesions on delayed imaging. No additional liver lesions were detected on further delayed imaging in the 286 patients. The other 24 patients with liver lesions identified on initial imaging still showed lesions on delayed and further delayed imaging. The analysis showed a significant difference in the percentage of colorectal cancer (66.7%) and liver lesions before the PET scan (50.0%) compared with unchanged results . Routine triple-time-point imaging did not improve the detection of liver metastases; however, it may be recommended in patients with colorectal cancer and liver lesions before the PET scan.Prior reports have demonstrated the improved ability of delayed fluorine-18 ( The detection of liver metastases is critical for cancer staging and disease management. Therefore, our newly opened positron emission tomography (PET) center used the current PET/CT scan to emphasize the accuracy of PET results, especially in the detection of liver metastases.18F) fluorodeoxyglucose (FDG) PET/CT is highly sensitive and specific for identifying patients with liver metastases [Fluorine-18  has been reported ,11,12,1318F-FDG administration [Different time points are set for initial and delayed imaging, the mean time intervals ranging from 50 to 90 min and from 100 to 120 min after stration ,12,13. Tstration .Taking advantage of previous reports in dual-time-point PET/CT imaging ,11,12,13The current PET/CT scan owns a large field-of-view coverage and the highest NEMA sensitivity. The study was designed to assess liver metastases detected only seen on delayed rather than on initial FDG PET/CT imaging to evaluate, whether such an additional examination time and radiation exposure are worthy. To our knowledge, no triple-time-point imaging has been reported.18F FDG PET/CT imaging. Demographic characteristics of the 310 patients and their imaging findings are shown in This retrospective study was approved by the institutional review board of our hospital. The need for written informed consent was waived . After excluding patients who were intolerant to longer examinations that did not involve delayed imaging, from March 2017 to April 2018, 310 patients with initial staging or the recurrence of various malignancies  who underwent FDG PET/CT scans for initial staging or the recurrence of various malignancies were analyzed using a current PET/CT. No liver metastatic lesions were detected before the scans. Hence, no lesions were previously treated by interventional radiofrequency (RF) procedures. All patients with cancer were surveyed with at least one liver imaging modality (ultrasound (US) imaging or enhanced CT) before We used the 16-section multidetector Discovery IQ PET/CT system with the five-ring configuration with a large, 26-cm field-of-view coverage and the highest NEMA  sensitivity in the industry, at up to 22 cps/kBq . The protocol of FDG PET/CT oncologic imaging started with low-dose CT without contrast medium from the vertex to the feet for attenuation correction of subsequent PET images and anatomic localization at approximately 60 min after the FDG injection. CT images were acquired using the 16-section multidetector scanner with an automatic exposure control system, performing 3D mA modulation according to the patient’s body size and attenuation level, ranging from 15 to 50 mA; the tube voltage was 120 kVp. Initial PET acquisition was performed in the same region, acquisition type 3D, 2 min/bed position from the vertex to the upper thigh and 1 min/bed position from the upper thigh to the feet ; orderedThen, delayed limited-area imaging including the liver was performed 100–120 min (mean 109 min) after the 370 ± 74 MBq FDG injection, and further delayed imaging focused on the liver was performed at 140–160 min (mean 151 min) with the same CT and PET parameters as the initial imaging, except for the 3- and 5-min/bed position on delayed and further delayed imaging for PET acquisition, respectively. In general, delayed imaging was acquired only in selected bodily regions based on patient history and findings on initial imaging, but had to include the liver, and further delayed imaging included only the liver.Two experienced nuclear medicine physicians with knowledge of prior findings detected liver lesions by visual interpretation in consensus using a Xeleris 3.0 workstation , separately. In case of disagreement, the cases were discussed and consensus was obtained.For proper interpretation, a thorough knowledge of the patient’s history was mandatory. Comparisons with other recent imaging modalities, such as CT, MRI, or ultrasound imaging, were performed for the diagnosis of the lesion, especially for the liver. If liver lesions were detected by any imaging modality including PET/CT, a final diagnosis of the presence of liver metastases or not would be pathologically confirmed if feasible under the patients’ informed consent, or would be correlated with the other liver imaging modalities. Enhanced CT has better diagnostic performance than US in the detection of liver metastases. Hence, we performed enhanced CT regardless of whether the detection of liver metastases on US was possible before the PET scan or not. MRI has higher sensitivity in detecting liver metastases than CT. Therefore, even if enhanced CT does not detect the metastatic liver lesions, they could be identified by enhanced MRI with Gadolinium or Primovist, a hepatospecific contrast medium . Alternatively, if no liver metastases were detected on enhanced CT before performing the PET scan, we reviewed the CT images. If CT images still yielded negative results, we then resorted to enhanced MRI. Therefore, enhanced CT or MRI served as the reference and gold standard in the detection of liver metastases in this research. Adequate management was adopted by the clinical physician. If no liver lesions were detected by any of the imaging modalities, the final diagnosis was no liver metastases.max) and tumor-to-normal liver ratio  were calculated and correlated with the visual interpretation. The SUVmax of the liver lesion on initial imaging was measured in the same region as that on delayed imaging.Patients with FDG-avid liver lesions detected only on delayed imaging, but not on initial imaging were the focus of this study (six patients in this study). The liver maximum standardized uptake value .A Among the 310 patients, all 24 patients with liver lesions  identified on initial imaging were still clearly seen on delayed and further delayed imaging, and there were 286 patients with no liver lesion on initial imaging . Of the max values among initial, delayed, and further delayed imaging in these six patients , negative predictive value (97.9% vs. 100%), and accuracy (98.1% vs. 100%).As per our preliminary data, six of the 286 patients (2.1%) had liver metastases identified only on delayed PET/CT imaging. The mere 2.1% of patients identified may not indicate the importance of delayed imaging, but when it detects liver metastases, it still plays a crucial role in individual disease management. Therefore, the value of delayed imaging should not be understated. The other 24 patients with liver lesions identified on initial imaging were still seen on delayed and further delayed images. Similar to a prior review article , such imIt may not be cost-effective or practical to perform delayed PET imaging, or even further delayed images, on every patient in PET centers with high patient throughput, and it may be more feasible to consider this approach in selected patients who have a higher likelihood of lesion detection only on delayed imaging. According to the findings in max values of these six patients with liver metastases identified only on delayed PET/CT imaging did not exhibit a continuous increase with time, but the corresponding TNRs continued to increase with time , negative predictive value (97.9% vs. 100%), and accuracy (98.1% vs. 100%); therefore, this imaging technology should be used in clinical practice to optimize diagnostic performance 0%, negat,12,13,16The application of further delayed liver imaging results in more radiation exposure to patients, but we used a current low-dose CT with a new technique  i.e., a.We used a triple-time-point protocol for our routine protocol at initial setup to see whether it could detect more patients with liver metastases. However, after this retrospective study, we changed our routine protocol to a dual-time-point protocol, except in patients with colorectal cancer and with liver lesions before the PET scan.In conclusion, as routine triple-time-point PET/CT imaging did not improve the detection of liver metastases, it can be recommended that in patients with colorectal cancer and with liver lesions before the PET scan, the use of a current PET/CT should be regarded as the most practical and cost-effective approach. Triple-time-point PET/CT imaging could confirm liver lesions and indicate the tendency of malignancy. However, more data are required for further clarifications in the future.


Among the latter group, ORs (95% CIs) were 3.05 (1.75–5.33) for PPI initiation after start of aspirin therapy, and 1.66 (0.63–4.36) for PPI initiation on/before start of aspirin therapy. For LGIB (n = 1428), ORs (95% CIs) were 0.98 (0.81–1.17) for >1 month PPI use and 1.12 (0.73–1.71) for ≤1 month PPI use. Among low-dose aspirin users, maintaining PPI use (>1 month) was associated with a significantly reduced UGIB risk. Neither short nor long-term PPI use affected LGIB risk.Estimates of the effect of proton pump inhibitors (PPIs) on risks of upper and lower gastrointestinal bleeding (UGIB and LGIB) among low-dose aspirin users in routine clinical practice are variable (UGIB) or lacking (LGIB). We aimed to establish these risks in the same observational study population. Using UK primary care data, we followed 199,049 new users of low-dose aspirin (75–300 mg/day) and matched non-users at start of follow-up to identify incident UGIB/LGIB cases. In nested case–control analyses, adjusted odds ratios (ORs) were calculated for concomitant PPI use vs. past (discontinued) PPI use among current low-dose aspirin users. For UGIB ( Low-dose aspirin is widely prescribed for the secondary prevention of cardiovascular disease (CVD) as this benefit is well established in outweighing the risk of major bleeding . The netIn population-based studies, estimates of the size of a PPI-induced reduction in upper gastrointestinal bleeding (UGIB) risk among users of low-dose aspirin have varied. Some studies have had limited power to produce precise estimates of effect, and definitions of drug use between studies have differed ,7,18,19.We used data from a population-based study of gastrointestinal bleeding set in UK primary care, using validated cases of both UGIB and LGIB. We used a matched cohort design with nested case–control analysis to quantify the effect that PPI use (vs. non-use) has on the risk of UGIB and LGIB among users of low-dose aspirin, using an appropriate reference group for PPI use that aimed to minimize confounding and selection bias. We analyzed both duration of PPI use (short- or long-term) and the timing of PPI initiation in relation to the start of low-dose aspirin therapy.We used The Health Improvement Network primary care database in the UK, which contains the electronic health records of approximately 6% of patients across the UK, who collectively are representative of the UK demographic ,21. ThisUse of low-dose aspirin among UGIB/LGIB cases and controls was categorized as follows: current use, when the supply of their most recent prescription lasted until (or over) the index date, or ended in the 30 days before the index date; past (discontinued) use, when low-dose aspirin use ended more than 1 month before the index date; never use, if there was no recorded low-dose aspirin prescription before the index date. Within current users of low-dose aspirin, PPI use was categorized using these same categories.Odds ratios (ORs) with 95% confidence intervals (CIs) were calculated using unconditional logistic regression adjusted for confounders using a stepwise forward approach. Patients with missing data on smoking, body mass index (BMI), alcohol consumption, estimated glomerular filtration rate or urban/rural residence were placed in a separate category ‘missing’ for that variable The fully adjusted model for the UGIB outcome was adjusted for the matching variables  and number of PCP visits in the year before the index date, alcohol consumption, smoking, history of UGIB, LGIB, unspecified gastrointestinal bleeding (GIB), peptic ulcer, pancreatic disease, polytherapy and use of clopidogrel, non-steroidal anti-inflammatory drugs (NSAIDs) and warfarin. The fully adjusted model for the LGIB outcome was adjusted for matching variables  and number of PCP visits in the year before the index date, alcohol consumption, smoking, body mass index, history of polyps, LGIB, unspecified GIB, peptic ulcer, gastro-esophageal reflux disease, inflammatory bowel disease, irritable bowel syndrome, polytherapy and use of clopidogrel, NSAIDs and warfarin. Polytherapy was defined as the number of different medications where a prescription lasted until/over the index date or lasted to within the 30 days before the index date. The main exposure of interest was current use of low-dose aspirin plus current use of a PPI  and, to minimize confounding from time-varying differences between PPI users and non-users that are difficult to control, the reference group was current use of low-dose aspirin plus past use of a PPI. To evaluate the likely effect of confounding by indication among individuals who had very recently started using a PPI , we categorized duration of PPI use as either ≤1 month or >1 month. To explore possible effects relating to the timing of PPI initiation among low-dose aspirin users, we further categorized PPI use for patients with ≤1 month duration of use according to whether the PPI was prescribed after the start of low-dose aspirin therapy or on/before the start of low-dose aspirin therapy. An additional analysis was also undertaken to evaluate the effect of concomitant PPI use (of any duration vs. past PPI use) among current users of dual antiplatelet therapy  on the risk of UGIB/LGIB. We did not perform any analyses for specific PPI agents. All analyses were conducted using STATA version 12.0.The main characteristics of UGIB/LGIB cases and controls are shown in The frequency distribution of low-dose aspirin and PPI use among UGIB cases and controls  is shown in The frequency distribution of PPI use among LGIB cases and controls  is shown in Owing to the fact that the majority of LGIB cases among low-dose aspirin users in UK primary care are referred to a specialist but not hospitalized (25), we performed a post-hoc analysis in which we repeated the main LGIB analysis according to whether LGIB cases were only referred (less severe cases) or were hospitalized (more severe cases). Odds ratios did not differ between these two groups of cases, with minimal change seen from the main estimate: OR 0.97 (95% CI: 79–1.18) for only referred cases, and 1.04 (95% CI: 0.78–1.39) for hospitalized cases.In our large population-based study, we have shown that among patients using low-dose aspirin for CVD prophylaxis, those who continued with PPI co-therapy for at least 1 month had a significant 31% reduced risk of UGIB compared with those who discontinued PPI use (past users). However, co-therapy with a PPI, whether in the short-term or long-term, had no effect on their risk of LGIB. The increased UGIB risk seen in our study soon after the start of PPI co-therapy  is most likely explained by protopathic bias due to the prescription of PPIs to treat prodromic symptoms of UGIB. Although interpreting the true effect of short-term PPI use is challenging in a non-randomized setting, results of our analyses separating out the timing of PPI initiation will help to elucidate the likely true effects of PPI co-prescription among low-dose aspirin users that are otherwise difficult to infer in observational studies when PPI use is grouped as single category.Our study provides valuable data on the effect that PPIs have on LGIB risk in low-dose aspirin users because previous data on this topic have been limited and findings have been mixed ,17. In lOur findings, which suggest that PPIs are effective as protective agents against low-dose aspirin-related UGIB, support several findings from previous reports on the topic ,7,18,19 A key strength of our study is that we provide evidence of the risk of both UGIB and LGIB in relation to PPI use among low-dose aspirin users from the same study population and time period. Also, the large sample from a population representative of the UK demographic enabled a large number of cases and controls to be obtained. Another strength is the use of a suitable comparison group to evaluate the effect of current PPI use. Because PPI use is a marker of comorbidity per se , the useOur results indicate that patients using low-dose aspirin who are co-prescribed a PPI have a significant 31% reduced risk of UGIB compared with those patients who discontinue PPI therapy, and do not have an increased risk of LGIB. Timely PPI co-prescription affords greater protection against UGIB compared with PPI co-therapy that is delayed.


Self-harm is a leading cause of morbidity in prisoners. Although a wide range of risk factors for self-harm in prisoners has been identified, the strength and consistency of effect sizes is uncertain. We aimed to synthesise evidence and assess the risk factors associated with self-harm inside prison.In this systematic review and meta-analysis, we searched four electronic databases  for observational studies on risk factors for self-harm in prisoners published from database inception to Oct 31, 2019, supplemented through correspondence with authors of studies. We included primary studies involving adults sampled from general prison populations who self-harmed in prison and a comparison group without self-harm in prison. We excluded studies with qualitative or ecological designs, those that reported on lifetime measures of self-harm or on selected samples of prisoners, and those with a comparison group that was not appropriate or not based on general prison populations. Data were extracted from the articles and requested from study authors. Our primary outcome was the risk of self-harm for risk factors in prisoners. We pooled effect sizes as odds ratios (OR) using random effects models for each risk factor examined in at least three distinct samples. We assessed study quality on the basis of the Newcastle-Ottawa Scale and examined between-study heterogeneity. The study protocol was registered with PROSPERO, CRD42018087915.I2=49%), lifetime history of suicidal ideation , and previous self-harm . Any current psychiatric diagnosis was also strongly associated with self-harm , particularly major depression  and borderline personality disorder . Prison-specific environmental risk factors for self-harm included solitary confinement , disciplinary infractions , and experiencing sexual or physical victimisation while in prison . Sociodemographic (OR range 1·5–2·5) and criminological (1·8–2·3) factors were only modestly associated with self-harm in prison. We did not find clear evidence of publication bias.We identified 35 independent studies from 20 countries comprising a total of 663 735 prisoners, of whom 24 978 (3·8%) had self-harmed in prison. Across the 40 risk factors examined, the strongest associations with self-harm in prison were found for suicide-related antecedents, including current or recent suicidal ideation (OR 13·8, 95% CI 8·6–22·1; The wide range of risk factors across clinical and custody-related domains underscores the need for a comprehensive, prison-wide approach towards preventing self-harm in prison. This approach should incorporate both population and targeted strategies, with multiagency collaboration between the services for mental health, social care, and criminal justice having a key role.Wellcome Trust. More than 11 million individuals are imprisoned worldwide.Previous research has examined a range of individual and environmental correlates of self-harm in prisoners, although findings are inconsistent across primary studies. ReviewsEvidence before this studyWe searched four databases  for systematic reviews of self-harm risk factors in adult prisoners published from database inception to Oct 31, 2019. The same keywords were used for each database search for self-harm, “(self-harm* OR suicid* OR attempt* OR NSSI OR self-injur* OR self-mutilat* OR self-destruct* OR poison* OR overdose)”, and prison “”. No language restrictions were set. We identified three systematic reviews with narrative summaries on risk factors for self-harm, non-suicidal self-injury, and near-lethal suicide attempt. These reviews reported that risk factors span many different individual and environmental domains, although there were many inconsistencies in the magnitude and direction of the effects. We did not identify any reviews that meta-analysed findings or evaluated the strength and consistency of risk factors for self-harm inside prison. We found one meta-analysis published in 2020 that examined the association between childhood maltreatment and suicide attempt in a population who has been in contact with the criminal justice system, but this meta-analysis included non-prisoners, juvenile offenders, and outcomes in the community.Added value of this studyThis meta-analysis synthesised data from almost 50 years of research examining risk factors for self-harm in over half a million prisoners. Although we identified many risk factors for self-harm across sociodemographic, criminological, custodial, clinical, and historical domains, the strongest associations were found for suicide-related exposures  and markers of psychiatric morbidity. Modifiable risk factors specific to prison include solitary confinement, disciplinary infractions, physical or sexual victimisation while in custody, and poor social support.Implications of all the available evidenceOur results show that risk factors associated with self-harm in prisoners include a range of potentially modifiable clinical, psychosocial, and environmental factors. These data emphasise the need for a whole-prison approach and multiagency collaboration in the prevention of self-harm.We did a systematic review and meta-analysis following the Preferred Reporting Items for Systematic Reviews and Meta-analyses guidelinesWe supplemented bibliographical database searches by hand-searching the citations and reference lists of relevant articles and previous systematic reviews.26LF screened the retrieved references for eligibility. We included primary studies that examined risk factors for self-harm in prison and met the following criteria: the study was cross-sectional, case-control, or cohort in design and included predominantly adult prisoners; the study was based on general prisoner populations ; the study included self-harm within prison as the outcome measure; and the study provided data for an appropriate control or comparison group of unselected prisoners who did not self-harm in prison.We excluded studies with qualitative or ecological designs, those that reported on lifetime measures of self-harm or outcomes other than self-harm  or on selected samples of prisoners , and those with a comparison group that was not appropriate  or not based on general prison populations.We contacted the authors of studies that did collect information on self-harm in prison but only reported prevalence,Data were independently extracted by two researchers (LF and Isabel Yoon). A standardised form was used to extract data and included information on study characteristics , sample details , outcomes , and risk factors. Extraction sheets for each study were crosschecked for consistency and any discordance was resolved by discussion between study authors. When the study characteristics were unclear, the corresponding authors of included papers were contacted. When multiple publications from the same population were available, information on risk factors was extracted from the investigation with the largest sample size. Data were only extracted from overlapping publications when a new risk factor was reported.As the reporting of effect sizes varied between studies, data were converted to a comparable measure for meta-analysis. Odds ratios (ORs) and their 95% CIs were extracted when reported or calculated from available data in the paper  by use of standard formulas.LF assessed all studies for risk of bias using the Newcastle-Ottawa Scale for cohort and case-control studies, with 9 points indicating high quality and low risk of bias.We grouped risk factors into five categories: sociodemographic, criminological, clinical, custodial, and historical. Three separate outcomes were identified: self-harm, suicide attempt, and non-suicidal self-injury. We have taken a broad definition of self-harm as any act of intentional self-poisoning or self-injury irrespective of the degree of suicidal intent or underlying motive,I2 statistic, which quantifies the percentage of variance across studies that can be attributed to true variation in effect sizes rather than sampling error as low (0–40%), moderate (30–60%), substantial (50–90%), and considerable (75–100%).To obtain a reliable estimate of pooled effect sizes, analyses were done only on risk factors examined in at least three distinct samples.The extent to which methodological variations across studies affected the association between risk factors and self-harm was examined by applying meta-regression models (by use of the metareg command). Specifically, univariate meta-regression analyses were done to explore sample size (n<median=0 and n≥median=1) and outcome definition  as possible sources of between-study heterogeneity for all risk factors. The presence of potential publication bias was assessed by examination of asymmetry in funnel plotsThe funder of the study had no role in study design, data collection, data analysis, data interpretation, or writing of the report. LF had full access to the data and all authors had final responsibility for the decision to submit for publication.Our systematic search of the literature identified 7436 unique records for screening, of which 454 (6·1%) full-text reports were examined for eligibility. We included 35 studies reported in 38 articles in the meta-analysis , compris67Included studies were done across 20 countries  and published from 1972 to 2020. Median sample size was 785 (IQR 142–2119), ranging from 60 to 263 794 prisoners. 20 (57%) of the 35 studies focused solely on either men (k=15) or women (k=5). In 15 studies, the sample included both male and female prisoners, with the mean proportion of women equalling 12·3% (SD 8·7), but only two studies provided data disaggregated by sex.In terms of study quality measured by the Newcastle-Ottawa Scale, of 9 possible points, the median score for the cohort studies was 8·5 (IQR 7–9) and the median score for the case-control studies was 6 (5–8). Of 8 possible points available in the modified Newcastle-Ottawa Scale, the median score for the cross-sectional studies was 6 (5–6). Overall, 18 (51%) of the 35 studies included were judged to be of high quality and four (11%) were categorised as being of low methodological quality .There were large variations in the sample sizes contributing to risk estimates. The largest samples were for sex (n=644 812) and violent offending (n=520 581). Only two risk factors (substance use disorder [n=766] and family history of suicide [n=382]) were calculated on the basis of a pooled sample of less than 1000 prisoners. Various sociodemographic factors were associated with self-harm in prison, with pooled ORs ranging from 1·5 to 2·5 . The thrClinical factors showed the strongest associations with self-harm compared with other risk factor domains  that might interact with custody-specific stressors  and thereby increase the likelihood of self-harming in prison. To address both predisposing and precipitating risk factors, the prevention of self-harm in custodial settings might require a comprehensive approach that comprises multilevel interventions, such as screening on reception, staff training ,Strengths of this investigation include a quantitative synthesis, large population numbers, and the inclusion of previously unavailable data. However, there are several limitations. First, despite our rigorous search strategy, which covered four major databases and scanning the reference lists of relevant studies and reviews, adding bibliographical indexes for criminal justice and global health might have identified additional work. Second, the strength of our risk estimates is likely to be overestimated because we did not account for confounding, and risk factors are unlikely to be independent of each other . Future work could provide more precise estimates by doing an individual participant meta-analysis, which would allow for the calculation of effect sizes adjusted in the same way. Third, reported associations could be due to reverse causality  and prospective studies are necessary to explore whether these risk factors predict self-harm during the course of imprisonment. Fourth, we did not consider the chronicity or frequency of self-harm because most primary studies reported a dichotomous self-harm outcome. Previous work suggests that repetition of self-harm is common, particularly in female prisoners,The accurate identification of individuals at risk of self-harm is challenging. Many of the identified risk factors, including suicidal ideation and psychiatric disorders,Given that previous self-harm is among the strongest predictors of future self-harmIn conclusion, a range of modifiable individual and prison-related factors increase the risk of self-harm in prisoners. Strategies to address these risk factors will potentially require interventions at all levels of the criminal justice system, including diverting people before prison, improvements to mental health care in prison, purposeful activities, and social support, and maintaining these approaches on release. Implementing these interventions will require a multisectorial approach across health, social care, and criminal justice.


Recent studies have shown an increase in open defecation and slippage of open defecation-free certified villages in Ethiopia, despite significant progress the country made on sanitation programs. Hence, realizing of existing facts, this study was conducted aiming at a critical review of available literature and to provide consolidated data showing the level of slippage and its associated factors in Ethiopia.Q and I2 test statistics were used to check heterogeneity among the studies. To negotiate heterogeneity from qualitative data, we used a mixed-method approach. The researchers also conducted a publication bias assessment and sensitivity analysis. A random effect meta-analysis was employed to determine the pooled estimates of open defecation free slippage rate with a 95% confidence interval (CI). The data analysis was performed using the CMA V.3 software program.Systematic literature searches were performed from four international databases. The search involved articles published from December 1, 2013, up to June 4, 2019. The Cochran’s After screening 1382 studies, 12 studies were finally included in this systematic review. The estimated pooled rate of open defecation-free slippage in Ethiopia was 15.9% (95% CI 12.9–19.4%). The main contributing factors for open defecation-free slippage were lack of technical support, financial constraints, low-quality building materials, improper program implementation, and lack of sanitation marketing.It was estimated that 1 out of 6 Ethiopian households engaged in open defecation after they have certified open defecation-free status, implying the low possibility of achieving sustainable development goals of 2030, which aims to ensure sanitation for all. Therefore, the government of Ethiopia and donors should better give special attention to the following options: (1) awareness for open defecation-free slippage, (2) launch a post-open defecation-free program, and (3) encourage research on pro-poor sustainable sanitation technologies.The online version contains supplementary material available at 10.1186/s13643-020-01511-6. Globally, around 0.9 billion people practice open field defecation. Recent reports showed a drop in the number of individuals practicing open defecation in many regions of the world. However, sub-Saharan African countries increased the number of people defecating in the open field from 204 million to 220 million .The reasons for increasing open defecation in Sub-Saharan Africa are high population growth and the slippage of open defecation-free (ODF)-certified communities, which refers to community member failure to keep fulfilling all open defecation free criteria. The investigation in African countries confirmed that the ODF slippage rate on the continent is 10–13% per year. Here, researchers defined the term ODF slippage based on sub-optimal latrine utilization and open field defecation .In Ethiopia, the sanitation program has given special attention since 1995 after the government incorporated public health in the National Constitution. Subsequently, the Ministry of Health developed the National Hygiene and Sanitation Strategy and National Hygiene and On-Site Sanitation Protocol in 2005 and 2006 consecutively . In 2006Community-Led Total Sanitation and Hygiene program is a better approach toward the reduction of open defecation practice and the achievement of the desired sanitation program. From 2011 to 2015, all Ethiopian administrative regions implemented a Community-Led Total Sanitation and Hygiene program covering more than 80% of the districts. According to the Ethiopian Ministry of Health report in 2015, 4657 kebeles (administrative villages) declared for open defecation-free status , 5.Similarly, reports from the Ethiopian Demographic Health Survey indicated a decreasing trend of open defecation: 81.9% in 2000, 61.9% in 2005, 38.3% in 2011, and 32.9% in 2016 , and appThe construction of more latrines does not ensure a drop in disease among people, especially children. However, open defecation-free status will reduce the occurrence of associated illnesses. Researchers confirmed that the prevalence of diarrhea was 7% in ODF villages compared to 38% in OD villages . SimilarRecently, increasing open defecation in Ethiopia was also due to open defecation-free slippage and high population growth. Most of Ethiopia’s ODF villages slipped back between 1 and 2 years of ODF certification . StudiesSo, this review aimed primarily to conduct a systematic review of the available literature of open defecation-free slippage rates in the country and to provide consolidated data showing the level of slippage and its associated factors. This review will provide comprehensive information contributing to policy reform on sanitation and hygiene-related health planning. It will also serve as baseline data for further study.We have employed a systematic review to estimate the pooled open defecation-free slippage rate and to determine the triggering factors of ODF slippage in Ethiopia. A mixed-method approach was used in this systematic review to assimilate data that have a qualitative nature (verbatim quotes and views) and quantitative data, whereas meta-analysis was also used to pool and present the data with a quantitative nature obtained from various studies . The mixSystematic reviews containing registered protocols were primarily explored to avoid duplication. We have also systematically explored articles and grey literature from international databases including Cochrane Library, PubMed, Google Scholar, and Science direct. The following search terms “Open Defecation OR Open Defecation-Free OR Open Defecation-Free Slippage OR Community-Led Total Sanitation AND Ethiopia” were used separately and along with the Boolean operators like “OR” or “AND” to get published articles and grey literature. Our search was performed from December 1, 2013, to June 4, 2019 (see S2 Table We used the following inclusion criteria for screening of the studies: (a) articles dealing with open defecation, open defecation free, open defecation-free slippage, and Community-Led Total Sanitation; (b) all kinds of study designs; (c) articles in which the outcome measures were an open field defecating households in open defecation-free villages or ODF slippage rate (event rate); (d) article in which ODF is achieved only through CLTSH program or intervention; (e) articles in which determining factors for open defecation-free slippage were described; (f) published full-text articles and unpublished grey literature; (g) articles published in the English language; (h) studies conducted in Ethiopia; and (i) articles published between December 1, 2013, and June 4, 2019. Articles which did not meet the above criteria were excluded from the study.Two researchers Thomas Ayalew Abebe (TAA) and Dr. Gudina Terefe Tucho (GTT) independently screened the titles and abstracts based on the inclusion criteria. They also removed duplicate articles. After removing duplicate articles, the full texts of the remaining articles were screened based on the inclusion criteria. The required information was extracted using a standardized data extraction format in Microsoft Excel and checked the data extraction process. These two researchers resolved their disagreements by consensus. The data extraction format included study characteristics such as the author name, region, study year, publication year, study design, study setting, sampling method, sample size, types of interventions to achieve ODF, ODF certification period, and open defecation-free slippage rate (event rate) Table .Table 1To examine the risk of bias or methodological quality, we analyzed each article by using Hoy et al.  tool forI2). The I2 ranges from 0 to 100%. Hence, one rule of thumb categorized the I2 values of 25%, 50%, and 75% as low, moderate, and high heterogeneity, respectively, with a p value less than 0.05 [The extracted data were entered in Microsoft Excel format and analyzed using the Comprehensive Meta-Analysis version 3 statistical software. The heterogeneity was detected by reviewing the tables describing the included studies, reviewing the forest plots, and reviewing statistical tests. Heterogeneity in this review refers to the percentage of variation across studies. It was quantified using the inverse variance  showed that 12 studies favored the risk of open defecation free slippage, whereas none of the studies favored ODF sustainability. If the event rate lies to the right of zero, it indicates open defecation-free slippage, and if it lies on zero, it shows open defecation-free status. However, here, zero has nothing to do with no effect. It only signifies no open defecation-free slippage or open defecation-free status. The data of open defecation-free slippage do not lie to the left side of zero because there was no negative effect size in this area of study. Open defecation free slippage data always skewed to the right or aligned on the centerline.The overall result from the forest plot (meta-view) favored open defecation-free slippage because the red diamond fell to the right of the centerline. The red diamond represents a summary of the pooled open defecation free slippage rate of the studies.I2 = 95%; P < 0.0001). Therefore, the researchers used random-effects meta-analysis for controlling heterogeneity. The significant magnitude of heterogeneity also suggested conducting subgroup analysis to identify the source . Heterogeneity was also high  came to be − 10.76 . This suggested the existence of publication bias that requires the trim-and-fill method to adjust the final pooled effect size. However, the trim-and-fill techniques indicated no studies were missing, and ODF slippage rate was unchanged based on a random-effects model  with the intercept (B0) of − 4.66 . Therefore, the reported ODF slippage rate (15.9%) is valid, and funnel plot asymmetry did not arise from publication bias.On visual inspection, the funnel plot was found to be asymmetric  difference.We accompanied a subgroup analysis based on the study area, sample size, and study design Table . When lip = 0.546).When restricting our analysis based on sample size (< 500 and ≥ 500), the sample size ≥ 500 subgroup had a higher open defecation-free slippage rate of 16.9% (95% CI 12.5–22.4%), whereas sample size < 500 had 14.8% (95% CI 11.0–19.7%); however, the difference between the groups was not statistically significant .Finally, when restricting our analysis by study design, the open defecation-free slippage rate was greater in the difference-in-difference design (25.40%) and lower in the survey (8.6%). In addition, heterogeneity between studies was statistically significant  of the heterogeneity between studies. The remaining unexplained heterogeneity was from the data . This slippage rate was higher than a study conducted in Nepal 3.5% , in Ghan.9–19.4%.The researchers of this review identified many factors for open defecation-free slippage that have emerged from qualitative content analysis and quantitative assessments and grouped these factors as socio-economic, technical, and policy-related issues.Nineteen percent of the selected studies indicated shared toilets as a contributing factor for open defecation-free slippage. Studies conducted in 13 African countries also confirmed that sharing latrines brought frustration even among family members. The poorest people without option and sharing toilet from their neighbors slipped back to open defecation . HoweverSeven (44%) of the research documents revealed that the reason for open defecation-free slippage was lack of money for maintenance. For this reason, they slipped back to open defecation or stopped using their latrines . Most poSome studies reported it as a contributing factor for open defecation. Households with filled latrines start open defecation because of inadequate space for reconstruction . A similThirteen percent (13%) of the reviewed studies showed a lack of freely available materials which was a barrier for sustaining open defecation-free status. The impact was high among poor people with limited free access to durable materials. Thus, latrine construction using short-lasting materials results in low-quality latrine and elevated open defecation-free slippage . A surveThirteen (81%) research documents reported a lack of technical support as contributing factors for open defecation-free slippage. The evaluation of Community-Led Total Sanitation and Hygiene effectiveness in 8 African countries similarly stated that interventions designed to advise on upgrading and improving sanitation facilities using local materials maintain open defecation-free status . SanitatLack of skill to construct safely managed latrine is another underlying factor for open defecation-free slippage. Usually, unskilled persons build a highly collapsible toilet. Latrine construction skills and technical advice help in the selection of better locally available materials and the correct latrine design. The unskilled person is unable to innovate good-quality latrine design. The few innovators available in the village were also unwilling to share their skills for others. Thus, most of the existing toilets were identical and unsanitary . People Poor quality of toilet was one of the contributing factors for open defecation-free slippage, as reported in 25% of the analyzed research. Poor-quality latrine pushes back users because of unaesthetic conditions (unpleasant smell or no privacy). Similarly, a study conducted in seven developing countries reported poor-quality latrine was a primary challenge for the sustainable use of a toilet in open defecation-free villages . SupportSoft collapsible soil was another contributing factor for frequent latrine collapse and open defecation. The study conducted in Mozambique illustrated that 60% of people did not rebuild their latrine in sandy soil . Latrinen = 5, 31%). Improper CLTSH triggering drove people to open defecation-free slippage. So, attention for the three elements of CLTS triggering  can have a long-lasting effect on sustained hygienic behavior [Improper ignition or triggering theme frequently appeared in five analyzed papers  of the analyzed documents. The unacceptable pressure on communities for early certification of ODF will contribute to slippage. Technology acceptability and changing behavior toward its utilization require a sufficient time of adoption. The ODF certification based on social pressure and solidarity has a long-term effect on ODF sustainability [Early certification of ODF appeared in 31%  of the research papers, the lack of enrollment comprehends all segments of the population in the sanitation system led to ODF slippage. Most programs ignored engagement, particularly disadvantaged people such as needy families, elders, children, disabled persons, women, and other vulnerable community members. Ignoring these groups of people affects the sustainable performance of open defecation-free status. An evaluation report in Sierra Leone depicted that children who continued open defecation were students who went to school exactly at the same time as CLTSH triggering or who did not attend the CLTSH triggering event [In 16%  will be difficult because of open defecation-free slippage. The planet has reduced open defecation radically since 2000, except for sub-Saharan Africa and Oceania. WHO and UNICEF reported that the factor for increment in open defecation was high population growth. However, the researchers of this review argue that population growth was not the only reason for open defecation increment, but the occurrence of open defecation-free slippage was another factor for the increment . If ODF Ethiopia is one of the top-piloted countries where open defecation tremendously decreased between the years 2000 and 2015. Even though the country has many most effective enabling environments, the reality depicts behavior change, and the implementation process through total sanitation and hygiene program is in back-and-forth progress because of emerging ODF slippage.Open defecation-free slippage will be an obstacle for achieving the Ethiopian Health Sector Development Plan (HSDP) in decreasing the percentage of households using latrines and the proportion of villages (kebeles) free of open defecation . MoreoveThe prevalence of waterborne diseases and neglected tropical diseases  is increasing in Ethiopia due to open defecation-free slippage. These diseases increase maternal and childhood morbidity and mortality, which in turn impacts the economic development of the country , 118–120Policy modification for equity and inclusion of poor and marginalized groups contributes to greater ODF sustainability. Therefore, it requires prioritization of the following activities at different levels.At a technical level, it is crucial to identify the best hygiene promotion approach and affordable sanitation technology and determine the minimum cost that will not discriminate the poor and marginalized groups. The strengthening of best approaches such as Community-Led Total Sanitation and Hygiene program and school WASH program will achieve sustained open defecation-free communities. The program like post-open defecation-free follow-up should be incorporated in the National Sanitation Program.At the government or donor level, the strategies and policies should better incorporate regular follow-up with measurable indicators. For example, including post-open defecation-free follow-up in the Community-Led Total Sanitation and Hygiene program will sustain the learned hygienic behavior. The government or donor should better consider the disadvantaged people who lack the resources  with acceptable subsidy programs because it will help them climb the sanitation ladder from a shared toilet to a private toilet or from a simple pit latrine to an improved sanitation facility. The increment of budget allocation for sanitation and hygiene programs also has a significant impact on program sustainability .At the researchers’ level, it is necessary to identify and test improved sanitation technology and estimate their cost. For instance, strategies that integrate Community-Led Total Sanitation and Hygiene with other low-cost technologies like ecological sanitation  have a significant impact on open defecation-free sustainability. It is common for rural communities to dig new pit after filled latrines, and finally, no reconstruction space or land left because of no tradition of reusing the pit. Furthermore, they have no idea that human excrement is used as a source of energy and organic fertilizer.Our findings indicate that applying these three alternatives helps eradicate open defecation or sustain hygienic behavior in the community. The existing National Sanitation and Hygiene strategy should include post-ODF program for the most straightforward implementation and for a better capacity to address ODF slippage. Finally, an appropriate sanitation technology selection, sanitation behavior remodeling, and assessing the levels of social belief help in sustaining open defecation-free status.Even though Ethiopia has adopted Sustainable Development Goal and reduced open defecation for decades, still the progress of improved sanitation facility is steady, and the ODF-certified villages were also retiring. It was estimated that one out of six Ethiopian households engaged in open defecation after they certified ODF status. Therefore, the increment of open defecation practice in Ethiopia was due to open defecation-free slippage besides population growth.The factors for open defecation-free slippage were lack of technical support, poor-quality construction material, lack of follow-up, financial constraints, improper implementation of Community-Led Total Sanitation and Hygiene program, demotivation due to the poor-quality latrine, demotivation using shared latrines, collapsible soil conditions, lack of freely available materials, lack of land, lack of involvement, and lack of promotion on sanitation marketing.Consequently, to accelerate the elimination of open defecation, the government of Ethiopia and its partners should better consider the level of slippage and its associated factor. Furthermore, there should be an integration of post-open defecation-free program and multi-sectoral coordination in the sanitation program. Finally, the encouragement of further research on pro-poor improved sanitation technology will help the community in climbing the sanitation ladder. Therefore, it will be easy to attain sustainable open defecation-free status or end open defecation by the end of 2030.Additional file 1. Table A. PRISMA 2009 Checklist for Open defecation free slippage assessment in Ethiopia, 2020.Additional file 2. Supplementary tables B–E.Additional file 3. Supplementary figures A–C.


This cross-sectional study evaluates whether hospital-specific adverse event rates are associated with hospital-specific risk-standardized 30-day episode-of-care Medicare expenditures for fee-for-service patients discharged with acute myocardial infarction, heart failure, or pneumonia. What is the association between 30-day episode-of-care expenditures and in-hospital adverse events?This cross-sectional study of 44 807 patients, which linked the 2011 to 2016 hospital-specific risk-standardized 30-day episode-of-care expenditure data from the Centers for Medicare & Medicaid Services and medical record–abstracted in-hospital adverse event data from the Medicare Patient Safety Monitoring System, found that hospitals with high adverse event rates were more likely to have high 30-day episode-of-care Medicare expenditures for patients discharged with acute myocardial infarction, heart failure, or pneumonia.This study suggests that hospitals with higher adverse event rates are more likely to have higher costs for acute myocardial infarction, heart failure, or pneumonia. Studies have shown that adverse events are associated with increasing inpatient care expenditures, but contemporary data on the association between expenditures and adverse events beyond inpatient care are limited.To evaluate whether hospital-specific adverse event rates are associated with hospital-specific risk-standardized 30-day episode-of-care Medicare expenditures for fee-for-service patients discharged with acute myocardial infarction (AMI), heart failure (HF), or pneumonia.This cross-sectional study used the 2011 to 2016 hospital-specific risk-standardized 30-day episode-of-care expenditure data from the Centers for Medicare & Medicaid Services and medical record–abstracted in-hospital adverse event data from the Medicare Patient Safety Monitoring System. The setting was acute care hospitals treating at least 25 Medicare fee-for-service patients for AMI, HF, or pneumonia in the United States. Participants were Medicare fee-for-service patients 65 years or older hospitalized for AMI, HF, or pneumonia included in the Medicare Patient Safety Monitoring System in 2011 to 2016. The dates of analysis were July 16, 2017, to May 21, 2018.Hospitals’ risk-standardized 30-day episode-of-care expenditures and the rate of occurrence of adverse events for which patients were at risk.The final study sample from 2194 unique hospitals included 44 807 patients  with a mean (SD) age of 79.4 (8.6) years, and 52.0% were women. The patients represented 84 766 exposures for AMI, 96 917 exposures for HF, and 109 641 exposures for pneumonia. Patient characteristics varied by condition but not by expenditure category. The mean (SD) risk-standardized expenditures were $22 985 ($1579) for AMI, $16 020 ($1416) for HF, and $16 355 ($1995) for pneumonia per hospitalization. The mean risk-standardized rates of occurrence of adverse events for which patients were at risk were 3.5%  for AMI, 2.5%  for HF, and 3.0%  for pneumonia. An increase by 1 percentage point in the rate of occurrence of adverse events was associated with an increase in risk-standardized expenditures of $103  for AMI, $100  for HF, and $152  for pneumonia per discharge.Hospitals with high adverse event rates were more likely to have high 30-day episode-of-care Medicare expenditures for patients discharged with AMI, HF, or pneumonia. However, few empirical studies have linked adverse events and expenditures across a large number of institutions.The US health care system is moving toward high-value care, with the goal of producing the best health outcomes at the lowest cost.18 and were largely focused on inpatient cost.27 Information is needed to examine the association between hospital performance on adverse events and hospital performance on episode-of-care expenditures within a standard period after admission in a contemporary and national cohort.A conceptual association between adverse events and expenditures could be that patients who have in-hospital adverse events may require additional expenditures to treat these adverse events. Such additional expenditures may also occur after discharge. Nevertheless, restricted by available data, previous studies were limited by the use of only a small number of measures28Accordingly, we sought to investigate the association at the hospital level between in-hospital adverse events and 30-day episode-of-care Medicare expenditures for Medicare fee-for-service patients with acute myocardial infarction (AMI), heart failure (HF), or pneumonia, 3 common conditions among older adults. The study used 2 unique national data sets, the hospital-specific Medicare 30-day episode-of-care expenditure data from the Centers for Medicare & Medicaid Services (CMS) and the adverse event data from the Medicare Patient Safety Monitoring System (MPSMS) to conduct this analysis. The 30-day Medicare episode-of-care expenditure data include all-source Medicare payments directly associated with care for individual services. It was the first database of its kind to be made available, and the MPSMS data represent the nation’s largest randomly selected hospital medical record–abstracted adverse event database. The setting was acute care hospitals treating at least 25 Medicare fee-for-service patients for AMI, HF, or pneumonia in the United States. Participants were Medicare fee-for-service patients 65 years or older hospitalized for AMI, HF, or pneumonia included in the MPSMS in 2011 to 2016. The dates of analysis were July 16, 2017, to May 21, 2018. In addition, we identified the best-performing hospitals in both expenditures and adverse events to represent high-value health care hospitals and assessed their characteristics.29 deemed that the requirement for informed consent could be waived for this cross-sectional study. The institutional review board at Solutions IRB reviewed the study protocol and granted a waiver of informed consent for the use of the deidentified database. This study followed the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) reporting guideline for observational studies.30The institutional review board at Solutions IRB31 The data include hospital-specific risk-standardized Medicare expenditures for episodes of care, starting with inpatient admission to a short-term acute care facility and extending 30 days after admission for Medicare fee-for-service patients . Because the classification of high-value care varies by condition, the range of this outcome is from 0 to 3, corresponding to none, 1, 2, and all 3 conditions in high-value care.42 We then performed a descriptive analysis to show patient and hospital characteristics and adverse event measurement across the 3 categories.Each hospital was classified into 1 of the following 3 mutually exclusive categories based on its risk-standardized 30-day expenditures: (1) low if the expenditures were in the lowest quartile (<25th percentile), (2) high if the expenditures were in the highest quartile (>75th percentile), and (3) average if otherwise.To evaluate the association between expenditures and adverse events at the hospital level, we fitted a linear regression model to link hospital-specific risk-standardized expenditures to the hospital-specific risk-standardized rate of occurrence of adverse events, with and without adjustment for hospital characteristics, including the hospital-specific Elixhauser Comorbidity Index score. The model was fitted for AMI, HF, and pneumonia separately.43 For each sub–data set and each hospital, we then calculated the risk-standardized rate of occurrence of adverse events described previously and fitted the above regressions to obtain a distribution and 95% CI for the estimate of the association between a hospital’s expenditures and the adverse event measure. To align with the CMS method for outcome measurements that restricts the analysis to hospitals with at least 25 discharges in the expenditure data, we conducted additional analyses by restricting the sample to hospitals with at least 25 adverse events for which patients were at risk over the study period.To address potential uncertainty in the estimates of expenditures and adverse events, we conducted bootstrapping analyses. Because the expenditure data were only available at the hospital level, parametric bootstrapping was used to generate 2000 random data points based on the hospital-specific point and interval estimates in the expenditure data. The inverse of variance of these bootstrapped data points was used to weight by their precision in the regression analyses described above. Because the MPSMS data were at the individual patient level, we used nonparametric bootstrapping with replacement to generate 2000 random data sets using the method developed for the CMS outcome measurements.Finally, we fitted a negative binomial regression model to assess hospital characteristics associated with high-value hospitals weighted by the hospital-specific number of exposures for which patients were at risk. Analyses were conducted using SAS, version 9.4, 64-bit (SAS Institute Inc).The final study sample based on linked CMS and MPSMS data across 2194 unique hospitals included 44 807 patients , with a mean (SD) age of 79.4 (8.6) years, and 52.0% were women. The patients represented 84 766 exposures for AMI, 96 917 exposures for HF, and 109 641 exposures for pneumonia. Patient characteristics varied by condition but not by expenditure category. The mean (SD) ages were 78.2 (8.7) years for AMI, 80.2 (8.5) years for HF, and 79.1 (8.6) years for pneumonia, and women accounted for 47.0%, 54.7%, and 52.3% for each condition, respectively . HospitaThe mean (SD) risk-standardized expenditures were $22 985 ($1579) for AMI, $16 020 ($1416) for HF, and $16 355 ($1995) for pneumonia per hospitalization  for AMI, 2.5%  for HF, and 3.0%  for pneumonia and varied by expenditure group . HospitaThe risk-standardized rate of occurrence of adverse events was associated with the risk-standardized expenditures for all 3 conditions  for AMI and $116  for HF per discharge. This association was reduced for pneumonia   for AMI, 189 of 2029 (9.3%) for HF, and 71 of 2060 (3.4%) for pneumonia, and they treated 5.6% of patients with AMI, 6.3% of patients with HF, and 5.9% of patients with pneumonia. Together, they represented 291 of 2194 unique hospitals (13.3%), of which 2 (0.7%) delivered high-value care for all 3 conditions, 38 (13.1%) for 2 conditions, and 251 (86.3%) for 1 condition. High-value hospital characteristics varied by condition . HospitaThis study used the hospital-specific risk-standardized rate of occurrence of adverse events as a proxy measurement of the hospital performance on adverse events. We found that the hospital performance on adverse events was associated with hospital-specific risk-standardized 30-day episode-of-care expenditures for patients with AMI, HF, or pneumonia. This finding suggests that investment in reducing adverse events may provide substantial savings in Medicare cost. Although the rationale to reduce adverse events goes far beyond economics, we believe that empirical data from across the country demonstrated the alignment between adverse events and cost.7 and were more likely to be readmitted,15 at least for AMI. The Office of Inspector General found that two-thirds of Medicare hospital costs associated with adverse events were the result of additional hospital stays necessitated within the same calendar month as the index hospitalization because of harm from the adverse event.16 Complications resulting from in-hospital adverse events may also cause additional adverse events after discharge, rendering these patients more likely to receive post–acute care services in skilled nursing facilities, home health care, and outpatient visits, as well as unplanned readmissions; consequently, such patients have higher 30-day episode-of-care risk-standardized expenditures compared with patients who do not develop an adverse event during their hospitalization.45 It is also possible that these patients were provided post–acute care services with higher rates of ambulatory care and follow-up with a condition-specific specialist after discharge, which are associated with additional expenditures.46There are several possible explanations for our findings. Patients who developed in-hospital adverse events probably required more care or were at increased risk of mortality26 Previous studies51 were restricted to in-hospital cost, but this study extends the cost from in-hospital to a 30-day standard period. For example, Zhan et al48 found that Medicare paid an extra $300 million in 2002 for 5 types of adverse events . Spector et al49 found that the occurrence of a hospital-acquired pressure ulcer was associated with an estimated $792 million in additional hospital costs that were incurred nationwide. Tsai et al50 found that patients who had major surgery at high-quality hospitals cost Medicare less than patients who had major surgery at low-quality institutions. Shamliyan and Kane51 found that hospitalizations associated with drug poisoning comprised 0.8% of all Medicare hospitalizations, with an annual hospital cost of $4 billion in 2008; in-hospital adverse drug events occurred during 5.3% of all Medicare hospitalizations. However, none of these studies captured expenditures for both inpatient and post–acute care services for AMI, HF, or pneumonia.The present study based on medical record–abstracted adverse event information was a large population-based investigation to assess the association between hospital performance on adverse events and 30-day expenditures for an episode of care for AMI, HF, or pneumonia in a contemporary cohort of Medicare beneficiaries in the United States. The use of risk-standardized 30-day payment data allowed us to capture costs not only during an index hospitalization but also immediately after discharge, a period in which substantial variation in Medicare expenditures exists predominantly because of differential use of post–acute care services.16 The reason is because these claims did not include diagnosis or procedure codes associated with the adverse events. Even if the claims included codes associated with the events, the codes often had no association with payments because the claims included other costly diagnoses or procedure codes that elevated the reimbursement to equivalent or higher amounts. Nevertheless, hospitals often must absorb the cost for these events. Researchers in Canada found savings of $9.1 million after implementing an infection prevention and control system that cost $6.7 million.52 Pettker et al53 reviewed liability claims at a single tertiary care teaching hospital for two 5-year periods (1998-2002 and 2003-2007) before and after implementing a safety program. They found that both liability claims (30 vs 14) and expenditures ($50.7 million vs $2.9 million) declined with the program.Reductions in adverse events often require investment in additional resources, which could increase a hospital’s overall budget and operating costs in the short term. However, from a long-term perspective, such an investment may reduce both Medicare expenditures and hospital costs, in addition to the primary objective of delivering safer care. The Office of Inspector General found that 84% of adverse events did not add to the Medicare payment for an inpatient stay.55 showed that adverse events frequently occur during the index hospitalization and adverse events that occur after hospital discharge are typically drug related. Restricted by the MPSMS data, we were unable to assess whether some of the measured adverse events have stronger associations with Medicare expenditures than others. It is possible that a proportion of the adverse events detected in the MPSMS may not be preventable, although each of the 21 in-hospital adverse event measures is characterized as being frequently preventable with the delivery of high-quality care. The study may also have underestimated the association between expenditures and adverse events because it is possible that some of the 21 adverse events may require care beyond a 30-day period. Limited by available expenditure data, we were unable to assess the expenditures from direct treatment of adverse events, and it is plausible that some expenditures may be associated with unmeasured confounding factors and that these expenditures may not be attributable to differences in adverse events. In addition, poor hospital performance on adverse events could be a marker of other systemic contributors and mechanisms, such as lower staffing ratios associated with care inefficiency and longer length of stay. Although the scope of this study constrained our ability to address these limitations in depth, future studies are warranted to elucidate them. Nevertheless, this study distinguishes itself by the breadth and standardization of events measured and its national scope.This study has limitations. We focused on adverse events that occurred during the index hospitalization and not after discharge; therefore, some events may have been missed. However, Forster et alThis study suggests that hospitals with poor performance on adverse events are likely to have high 30-day expenditures for AMI, HF, and pneumonia. These findings strengthen the evidence that adverse events may reflect the quality of hospital care and their reduction may be used as a mechanism for decreasing Medicare expenditures.