In clinical studies, it has been observed that esophageal cancer (EC) patient prognosis can be very different even for those patients with tumors of the same TNM stage. Tumor length has been analysed as a possible independent prognostic factor in many studies, but no unanimous conclusion has been reached. Therefore, this review used a meta-analysis to evaluate the association between tumor length and prognosis in EC patients.A systematic search for relevant articles was performed in PubMed, Web of Science, and Embase. Hazard ratios (HRs) and 95% confidence intervals (CIs) were used as effective measures to estimate the correlation between tumor length and prognosis, including overall survival, disease-free survival, progression-free survival, disease-specific survival, and cancer-specific survival. STATA 15.0 software was used to perform the meta-analysis and the data synthesis.p < .001) and disease-free survival (DFS)  in EC patients. Subgroup analyses also suggested a significant correlation between long tumors and poor OS. Sensitivity analysis and publication bias evaluation confirmed the reliability and stability of the results. Similar results were obtained in the analyses of progression-free survival (PFS), disease-specific survival (DSS), and cancer-specific survival (CSS).Finally, 41 articles with 28,973 patients were included in our study. The comprehensive statistical results showed that long tumors are an independent prognostic parameter associated with poor overall survival (OS) (HR = 1.30; 95% CI: 1.21–1.40, The results of this meta-analysis showed that long tumors were related to poor OS, DFS, PFS, DSS and CSS in EC patients. Tumor length might be an important predictor of prognosis in EC patients, and it can be used as an independent staging index. Further well-designed and large-scale prospective clinical studies are needed to confirm these findings. Esophageal cancer (EC) is one of the most common and fatal cancers of the digestive system in the world. In the United States, more than 18,000 people are diagnosed with EC every year, and more than 15,000 people die of EC. Most patients die within 1 year of diagnosis , 3. AlthThe current classification criteria for EC patients include histological grading, primary tumor (T), lymph node metastasis (N), and metastasis (M). Tumor length was not included as a prognostic indicator . In receAll data included in this study are publicly available in PubMed, Embase, and Web of Science. Our meta-analysis was based on the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) 2009 guidelines . This syPatients with EC were histopathologically confirmed, including SCC and AC;Studies explored the relationship between the tumor length and prognosis of EC;The hazard ratios (HRs) and 95% CIs between tumor length and survival outcomes were reported.Articles were included if they met the following criteria:They were abstracts, case reports, reviews, letters, or editorials;Studies were not published in English;Studies had overlapping or repeating data;Studies did not present the cut-off value for tumor length.Two researchers (ZW and YJ) independently reviewed articles that met the inclusion criteria. Any disagreements were resolved by discussion or arbitrator (WX). The HRs and 95% CIs were extracted and pooled according to different prognoses (OS or DFS/PFS/DSS/CSS) to analyse the relationship between tumor length and prognosis. All HRs were extracted from the multivariate analysis. The data were obtained directly from the included articles or from simple indirect data calculations . In addiP < .05 for the Q test or I2 > 50% for the I2 test indicated significant heterogeneity in the literature, and the random-effects model (DerSimonian-Laird method) was adopted [P < .05 was considered statistically significant.All statistical analyses were performed using STATA software version 15.1 . The pooled HRs and 95% CIs were used to assess the prognostic role of tumor length in EC patients. A pooled HR > 1 indicated a worse prognosis in EC patients with long tumors.  adopted ; otherwi adopted . Subgrou adopted , 19. TheBased on the search strategy mentioned above, a total of 1110 articles were searched from PubMed, Embase and Web of Science. Finally, 41 studies published between 1994 and 2019 with a total of 28,973 patients were included in our meta-analysis according to the inclusion and exclusion criteria . The spe2 groups were measured from postoperative pathological specimens, and the data from the chemoradiotherapy (CRT) group and Surg3 group were measured by endoscopy or computed tomography (CT) before treatment. The Surg1 group did not clearly describe the timing of measurement according to the different treatments. Because of the irregular shapes of the tumors, the actual size of them is difficult to measure. Most studies only analyzed the relationship between tumor length and patient prognosis, but did not discuss the effect of width. So, we default that the length of the tumor is proportional to its size, and combined the results of included studies on this basis. Two studies reported only relative risk (RR) data [Table RR) data , 42, andI2 = 66.8%; ph < .001), the pooled HR and 95% CI were calculated by a random-effects model. The pooled HR of 1.301 (95% CI: 1.210–1.399) suggests that long tumors are associated with poor OS . Due to obvious heterogeneity , the random-effects model was used to calculate the pooled HR and its 95% CI . The data suggest that long tumors were associated with shorter PFS.No significant heterogeneity was found among the studies that evaluated PFS, so a fixed-effects model was used to calculate the pooled HR and 95% CI . The pooled HR and 95% CI  indicated that long tumors were also related to poor CSS/DSS.Two studies reported preoperative tumor length and CSS/DSS data for EC. There was no significant heterogeneity (2 and others), median age (< 60 and ≥ 60), cut-off value (< 5 and ≥ 5), sample size (< 200 and ≥ 200), percentage of patients with LN metastasis (< 50% and ≥ 50%), percentage of patients with TNM stage III/IV classification (< 50% and ≥ 50%) and HR analysis method (MV and UV) were performed to explore the potential sources of heterogeneity for the pooled OS results. Because the cut-off values of tumor length were different among the included studies, we performed subgroup analysis according to different cut-off values. Because 40% of articles chose 5 cm as the cut-off value for tumor length, we used this cut-off value as the boundary. For cut-off value ≥5, the pooled HR was 1.259 . For cut-off value < 5, the pooled HR was 1.322 . This suggests that we may be able to define 5 cm as a standard cut-off value and recommend it to other researchers to reduce heterogeneity between different studies. The results of subgroup analyses showed that the different classification methods had no obvious influence on HR , histology , treatment   Fig. . TherefoEC is one of the most lethal malignant tumors in the world, with a 5-year survival rate of less than approximately 20% . With thI2 = 84.10%, ph < .001) [I2 = 67.90%, ph < .05) [I2 = 78.50%, ph < .01) [2 = 81.1%, ph < .001) [Therefore, we conducted this meta-analysis to assess the prognostic significance of tumor length in EC patients. Tumor size has been used in staging assessment systems for lung and thyroid cancers , 59. Man < .001) , 50–55. h < .05) , 52 and h < .01) , 55. Thih < .01) , 48. In  < .001) , 33, 50. < .001)  considerThere were several limitations in this meta-analysis that need to be addressed. First, our study only included research published in English, which may cause publication bias in this study. Second, the number of articles describing tumor length and DSS/PFS/DSS/CSS was insufficient, and the reference value was limited. Third, there were no corresponding data to analyse the relationship between tumor length and prognosis in different tumor locations. More importantly, there were many confounding factors, such as different resources of patients, pathological types, treatment strategies, durations of follow-up, sample sizes, proportions of lymph node metastasis, TNM stages and cut-off values of tumor length, which will increase the heterogeneity of this study. Although we conducted subgroup analyses and sensitivity analyses on the extracted data, these analyses could not fully explain the heterogeneity. Significant heterogeneity may affect the reference value of the meta-analysis results. We also assessed Begg’s funnel plot and performed Egger’s test to evaluate publication bias, and the results indicated that 12 articles on the relationship between tumor length and OS in EC have not been published. Though we adjusted the pooled result with the trim-and-fill method, the existence of significant publication bias suggests that we may have deficiencies in the selection and inclusion of literature. All of these limitations may hinder the application of the results in clinical work. More studies and larger sample size meta-analyses are needed to correct these limitations and improve the accuracy of the results.The tumor length of EC is widely measured in clinical work, but it is not often used as an indicator to evaluate prognosis and select treatment. Our results show that long tumors are related to poor survival outcomes in EC and can be considered an effective prognostic factor. This index can be used for the risk stratification of patients and is a promising prognostic index for clinical decision-making regarding EC treatment, especially for early-stage patients. However, due to the limitations listed above and the lack of more detailed case-control studies, more well-designed and large-scale studies are needed to confirm this conclusion in the future.


For discriminant analysis, relapse or success status was added to each patient’s T0 data.To evaluate morphologic differences between class III malocclusion success and failure treatment subjects in order to identify which variables are more predictive for long-term stability in early orthopedic treatment. In this retrospective study, 31 patients were enrolled from the Department of Orthodontics (Rome Tor Vergata). Inclusion criteria were as follows: white ancestry, class III malocclusion, mixed dentition, cervical stage (CS) 1-2, no pseudo-class III. Pre-treatment radiographic and cast records were collected. Each patient underwent rapid maxillary expansion/facial mask/bite block (RME/FM/BB) orthopedic treatment until correction. At T1 , records were recollected. According to treatment stability, relapse group  and success group  were identified. Sagittal and vertical cephalometric and digital cast measurements were performed. Student’s p = 0.0266), while at T1 the upper anterior length was shorter than SG (p = 0.0028). Between T1 and T0, both groups showed larger upper anterior and posterior transversal widths. SG had greater upper anterior (p = 0.0066) and posterior (p = 0.449) sagittal length. RG presented larger lower anterior (p = 0.0012) and posterior (p = 0.0002) transversal widths, while there were no differences in SG lower arch. Discriminant analysis provided two predictive variables with an accuracy of 80.6%: upper anterior length and upper posterior length.At T0, RG showed larger upper anterior transversal width (A shorter and wider maxilla could be a predisposing factor for relapse and failure of the early orthopedic treatment of class III malocclusion patients. The absence of mandibular changes could be predictable for treatment success. Treatment stability in class III malocclusion is a topic of crucial importance, as many factors are involved in prognosis and in long-term response. The etiology is wide-ranging and complex : A synerIn 1966, Tweed –24.Only few studies analyzed the predictive variables of treatment success in class III malocclusion on dental arches. In 1997, Franchi et al.  evaluateTo our knowledge, no studies assessed the presence of predictive stability variables in dental arches with three-dimensional analysis after early orthopedic therapy. The most popular early orthopedic treatment protocol for class III malocclusion is the combination of rapid maxillary expansion (RME) followed by maxillary protraction with facemask (FM) .For this reason, the aim of this study was to evaluate dental arches morphology in class III malocclusion subjects and to identify their morphologic differences by comparing a success and a failure treatment group after early orthopedic therapy.This project was approved by the Ethical Committee of the University of Rome Tor Vergata (Protocol number: 201/19), and informed consent was obtained from the patients’ parents.In this retrospective study, a sample of 31 patients , who were treated consecutively with RME/FM protocol and with bite block (BB) from 2005 to 2014, was enrolled from the Department of Orthodontics at the University of Rome Tor Vergata. Inclusion criteria were white ancestry, class III skeletal malocclusion  characterized by an anterior crossbite or edge-to-edge incisal relationship, early mixed dentition phase, skeletal maturation between cervical stage (CS) CS1-CS2 evaluated by cervical stage valuation (CSV) method as described by Baccetti et al. , no pseuAt pre-treatment phase (T0), radiographic records were acquired  and upper and lower dental casts were collected.Every patient underwent the same treatment protocol. A bonded rapid maxillary expander (RME) with vestibular hooks for facial mask (FM) was placed on first upper molars. The screw was activated one turn a day until transversal overcorrection achievement. At the end of active expansion, FM was given for maxillary protraction with extra-oral elastics, delivering 400–500 g of force per side, with about 30 degrees of downward inclination relative to the occlusal plane and patients were instructed to wear the mask 14-16 h per day (night-time included). Moreover, to control the vertical dimension an acrylic bite-block appliance (BB) with occlusal lift planes was used  and eachAt the end of this first phase of treatment, all patients reached the class III malocclusion orthopedic correction, and they underwent routine recalls to follow-up treatment stability.These patients were seen every 6 months until they reached a skeletal cervical maturation CS4 with all their permanent teeth erupted except for the third molars (mean age 14.5 years ± 5 months). Before starting the second phase of treatment (T1), radiographs and dental casts were acquired to assess orthopedic treatment stability, checking molar, and canine dental class, as reported by Wendl et al. , sagittaAccording to the treatment stability of early orthopedic therapy, two groups were identified: relapse group (RG) and success group (SG). RG  presented at T1 skeletal class III malocclusion , class III molar and canine relationship, unfavorable esthetical characteristic and needed for a new phase of therapy to achieve the correct occlusal relationship; on the contrary, SG  presented class I occlusion and good esthetical characteristics.On lateral cephalograms, linear measurements were performed to diagnose the class III skeletal malocclusion and evaluate patients’ divergence  Table  Fig. 1)1). To evIn a pilot study, 10 patients were used to calculate the reproducibility and the sample size which indicated the need for approximately 22 patients to estimate the upper anterior transversal width with a 95% confidence interval (CI); a minimum difference of 2.5 mm and a standard deviation (SD) of 2.5 mm, with a power of 80%.t test was used to compare the two measurements . In the presence of normally distributed data, descriptive statistics were calculated for each measurement in each group and significant between-group differences were tested with the independent sample Student’s t test. The level of significance was set at 5%.To determinate the method accuracy, measurements on digital dental casts were performed by one trained examiner with an experience of 5 years (FCDR) and repeated by the same approximately 15 days after. A paired Relapse or success status was added to preliminary patient treatment data. The measurements were examined using discriminant analysis, run under the IBM statistical package for the social sciences . Discriminant analysis is a statistic model, specifically designed to widely separate two groups of subjects taken from the same population . For thiNo systematic error was found.ANB = −0.878°; Wits = −6.306 mm) with a normodivergent pattern . According to treatment stability, two groups were identified, and intra-groups and inter-groups analysis were performed. At T0, no significant statistical differences were present between the two groups for vertical and sagittal cephalometric measurements. At T1, the cephalometric analysis showed a hyperdivergent pattern  with reduced values of ANB (1.3° ± 1°) and Wits (−11 mm ± 3.7 mm) in RG compared to SG , which resulted larger in RG (32.02 mm ± 2.83 mm) than SG (29.2 mm ± 3.89 mm) Table . At T1, p = 0.0012) and lower intermolar width (p = 0.0002) increased at T1. On the contrary, SG showed a statistically significant improvement in the upper anterior sagittal length (p = 0.0066) and in the upper posterior sagittal length (p = 0.0449). The lower arch showed no statistically significant changes.Examining the intra-group differences T1-T0 Table , both grUpper posterior sagittal length Upper anterior sagittal length Table These variables maximized the Mahalanobis distance between the two groups, with a Wilks’ lambda of 0.616 . As shown in Table The aim of our study was to assess the morphologic dental arches differences in patients with class III malocclusion by comparing a success and a failure treatment group after early orthopedic therapy (RME/FM/BB). Treatment of class III malocclusion is a challenge in orthodontics, as the result of failure is relatively frequent . TherefoIn this retrospective study, a sample of 31 patients with class III malocclusion was collected. According to the treatment stability of their early orthopedic therapy (RME/FM/BB) , they weEvaluating the dental arches morphology, the T1-T0 comparison showed both in RG and SG statistically significant increases in the maxillary width measurements. This result is due to the RME treatment, which separated the midpalate suture in a “V”-shaped pattern , 35 and Focusing on the mandibular arch, RG showed that lower intercanine and intermolar width increased significantly. In SG, there were no mandibular transversal or sagittal changes. These results agree with the discriminant analysis of Franchi et al. , who assAt T0, RG showed a greater upper intercanine width when compared to SG, while no significant differences were found in the upper intermolar width. This result disagrees with Wendl’s findings , which aTo evaluate which morphologic characteristic of the maxillary and mandibular arches could be used as stability predictive factors of early orthopedic treatment of class III malocclusion, a discriminant analysis was made. Other studies performed discriminant analysis , 24, 31,Our predictive model aimed to identify good or bad responders to interceptive class III malocclusion treatment (RME/FM/BB) by using dental arches morphology analysis in mixed dentition. Discriminant analysis found two predictive variables of relapse: upper posterior sagittal length and upper anterior sagittal length.Our results show that before treatment, RG subjects have a shorter and wider maxilla. At the end of the treatment, RG presents no significant improvement on the sagittal length if compared to SG, and treatment relapse in RG is further complicated by the transversal increase in mandibular arch.Our analysis does not imply that our model is able to classify surgical and non-surgical cases, yet it could help to identify eventually bad responders to therapy. A shorter and wider maxilla could be a predisposing factor for long-term relapse and failure.The main limitations of this study are its retrospective nature and the small sample size due to the low prevalence of class III malocclusion in Caucasian ancestry  and to Dental arches analysis is a useful diagnostic tool of stability prediction in the early interceptive class III malocclusion treatment and represents a method to predict relapse.Anterior and posterior maxillary sagittal lengths are two predictors of relapse in early orthopedic class III malocclusion treatment.A shorter and wider maxilla could be a predisposing factor for relapse and failure in early orthopedic treatment of class III malocclusion patients.The relapse patients show a statistically significant increase in the transverse dimension of the mandible. The absence of changes in the mandibular arch could be a predictive factor of treatment success.


Serum free fatty acid (FFA) concentrations are associated with coronary heart disease and diabetes mellitus (DM). Few studies focused on the relationship between serum FFA levels and coronary artery calcification (CAC).This was a retrospective, single-centered study recruiting patients underwent FFA quantification, coronary angiography and intravascular ultrasound (IVUS). CAC severity was assessed with the maximum calcific angle (arc) of the calcified plaque scanned by IVUS. Patients with an arc ≥ 180° were classified into the severe CAC (SCAC) group, and those with an arc < 180° were classified into the non-SCAC group. Clinical characteristics, serum indices were compared between 2 groups. Logistic regression, receiver operating characteristic (ROC) curves and area under the curves (AUC) were performed.p < 0.001). Logistic regression revealed that serum FFAs were independently associated with SCAC after adjusting for confounding factors in the whole cohort , the non-DM group  and the DM group . ROC analysis revealed a serum FFA AUC of 0.695  in the whole population. The diagnostic predictability was augmented  in the DM group and decreased  in the non-DM group.Totally, 426 patients with coronary artery disease were consecutively included. Serum FFA levels were significantly higher in the SCAC group than non-SCAC group (6.62 ± 2.17 vs. 5.13 ± 1.73 mmol/dl, Serum FFA levels were independently associated with SCAC, and could have some predictive capacity for SCAC. The association was strongest in the DM group.The online version contains supplementary material available at 10.1186/s12872-021-02152-w. Coronary artery calcification (CAC) has been traditionally recognized as a common complication in aging patients, and those with diabetes mellitus (DM) or chronic kidney disease (CKD) , 2. CAC Serum free fatty acids (FFAs) are one of the sources of energy in the body. They were found to be associated with insulin resistance and the development of DM –13. PrevTherefore, we attempted to explore the potential relationship between serum FFA levels and CAC in the present study.2 or who underwent hemodialysis; (7) severe hepatic dysfunction; (8) malignant tumor; or (9) patients with poor quality IVUS imaging or lack of laboratory test results. All patients gave written informed consent, and the study protocol was approved by the Institution's Human Investigation Committee. Procedures were performed in accordance with the Declaration of Helsinki.This was a single-center, retrospective study conducted at the Shanghai Ninth People's Hospital affiliated with Shanghai Jiao Tong University School of Medicine from February 2017 to February 2020. The enrollment criteria were as follows: (1) patients aged ≥ 18 years, and (2) patients who underwent CAG and IVUS for a de novo lesion and were diagnosed with CAD. The exclusion criteria were as follows: (1) previous PCI or coronary artery bypass grafting (CABG) on the target vessel; (2) patients with a lesion of chronic total occlusion; (3) moderate or severe cardiac valve disease; (4) NYHA class 3–4 heart failure; (5) type 1 DM (T1DM); (6) patients with a declined renal function of eGFR ≤ 30 ml/min/1.73 m2 (kg/m2). Blood pressure was measured on admission. Blood samples were routinely collected in the morning after overnight fasting when hospitalized, including lipid profiles (serum FFA was part of the fasting lipid profiles), kidney function, HbA1c et al. Estimated glomerular filtration rate (eGFR) were calculated according to CKD-EPI equation.Detailed medical history of the patients was collected after hospitalization. Weight and height were recorded and body mass index (BMI) was calculated by weight/heightThe CAG procedures were performed according to generally accepted guidelines and routines , 28. TheAll IVUS data were stored in DVDs and analyzed offline according to the criteria of the American College of Cardiology Clinical Expert Consensus Document on Standards for Acquisition, Measurement and Reporting of Intravascular Studies . AnalyseCoronary calcium was defined as a brighter plaque than adventitia with acoustic shadowing. The arc of the calcified plaque in the lesion was measured. Calcium length was determined as the length of the calcified plaque with an arc. According to the arc value, the cohort was divided into two groups: (1) Patients with an arc ≥ 180° were classified into the SCAC group, and those with a calcified arc < 180° or no calcium were classified into the non-SCAC group.p-value < 0.10 in single factor analysis. Receiver operating characteristic (ROC) curves were constructed to assess the diagnostic value of FFAs according to the area under the curve (AUC). All p-values and confidence intervals (CI) were two-sided, and p < 0.05 was considered statistically significant.Statistical analysis was performed with SPSS version 20.0 . Continuous variables are expressed as the Mean ± SD for normally distributed variables. Categorical data are presented as frequencies and proportions. Continuous variables were assessed using unpaired Student's t-tests, while categorical variables were compared using chi-square tests. Pearson’s correlation analyses between serum FFAs, and clinical and IVUS data were performed. Binary logistic regressions were used to calculate the correlations between the confounders and SCAC. Parameters were considered potential confounders if associations were found with a p < 0.001), more likely to have a history of DM , elevated HbA1c , increased pulse pressure  and decreased renal function . Serum FFA levels were significantly increased in the SCAC group . No significant differences in clinical presentation, serum cholesterol levels, total triglyceride levels, or use of drugs were observed between the two groups  were included. Table p < 0.001), a smaller MLA-LA  and a higher plaque burden  than the non-SCAC patients. Calcium length  and calcium arc  were also increased in the SCAC group (Table The distribution of SCAC in the coronaries was similar to that in the non-SCAC group Table . There wp = 0.001), lesion length , calcium length , calcium arc  and HbA1c . No positive relationship was found between serum FFAs and BMI  ,  and calcium parameters , except for a weak correlation between high density lipoprotein and calcium length , with a 41.4% increased odds of SCAC for every 1 mmol/dl increase in serum FFA. Model 2 revealed that in the non-DM group (n = 302), the relationship was weakened . As for the DM group (n = 124), the association was the strongest , with a 93.9% increased odds of SCAC for every 1 mmol/dl increase in serum FFA after adjusting for traditional cardiovascular risk factors. Serum FFA levels in the DM group were higher compared with the non-DM group  in the total population. In the non-DM group, the AUC was 0.649 . The AUC value (0.775) was the highest in the DM group .The predictive ability of serum FFAs for SCAC was assessed by ROC curve analysis Fig. . SimilarThe main findings of the present study were as follows: (1) Serum FFAs were independently associated with SCAC, and the relationship was enhanced in the DM group. (2) DM patients had a higher risk of elevated FFAs and SCAC. (3) Serum FFA levels may have some predictive capacity for SCAC, especially in the DM subgroup.The relationship between serum FFA levels and arterial calcification has been discussed in a few studies. Brooder MR et al.  found thSome previous studies examined the extent of CAC with CT , 26. In DM is correlated with CAC extent . Our datIn this study, in addition to FFAs, pulse pressure (PP) and age also served as predictors of SCAC, which was in line with previous studies , 36. CKDThis study had some limitations. First, this was a retrospective, single-center study with a relatively small sample size, and we focused only on the most severe coronary vessel without evaluation of the other coronaries. Second, we collected only serum fasting FFAs without postprandial FFA levels or various components of serum FFAs.Our present study indicated that serum FFAs may have some potential for predicting the severity of CAC, especially in DM patients. More prospective studies with large sample size are needed to further demonstrate the predictive value of FFA in CAC.Additional file 1: Table S1. Baseline clinical characteristics between non-diabetic (Non-DM) and diabetic (DM) groups.


Brassica napus ssp. napobrassica (L.) Hanelt) varies in root and leaf shape and colour, flesh colour, foliage growth habits, maturity date, seed quality parameters, disease resistance and other traits. Despite these morphological differences, no in-depth molecular analyses of genetic diversity have been conducted in this crop. Understanding this diversity is important for conservation and broadening the use of this resource.Rutabaga or swede , minor allele frequency (MAF) and mean expected heterozygosity (FST) and analysis of molecular variance (AMOVA) indicated that the rutabaga accessions from Norway, Sweden, Finland and Denmark were not genetically different from each other. In contrast, accessions from these countries were significantly different from the accessions from Iceland (P < 0.05). Bayesian analysis with the software STRUCTURE placed 66.9% of the rutabaga accessions into three to four clusters, while the remaining 33.1% constituted admixtures. Three multivariate analyses: principal coordinate analysis (PCoA), the unweighted pair group method with arithmetic mean (UPGMA) and neighbour-joining (NJ) clustering methods grouped the 124 accessions into four to six subgroups.This study investigated the genetic diversity within and among 124 rutabaga accessions from five Nordic countries  using a 15 K single nucleotide polymorphism (SNP) Overall, the correlation of the accessions with their geographic origin was very low, except for the accessions from Iceland. Thus, Icelandic rutabaga accessions can offer valuable germplasm for crop improvement.The online version contains supplementary material available at 10.1186/s12864-021-07762-4. Brassica napus ssp. napobrassica (L.) Hanelt, called ‘rutabagge’ in Sweden, ‘rutabaga’ in the USA and Canada, and ‘swede’ in the UK, New Zealand and Australia, is a cool-weather root crop thought to have been derived from the natural or spontaneous hybridization between B. rapa (turnip) and B. oleracea  [or kale) . Rutabagor kale) , 3. Neveor kale)  and was or kale) . Therefo1, B2, potassium, calcium, magnesium and iron [Rutabagas are grown for use as a table vegetable and as fodder for animals . The rooand iron , 6. Likeand iron . The leaand iron , 9. Howeand iron . Rutabagand iron , 11. Breand iron –15. Quanand iron .Genetic variation in plants is a key pillar of biodiversity and provides the resources for the development of new and improved cultivars with desirable characteristics . In addiB. napus [B. napus [In contrast, SNPs arise because of point mutations and hence most SNPs are biallelic, which leads to greater accuracy in genotyping; these markers also offer the advantage of co-dominance. In addition, SNP-based systems lend themselves to automation, and hence a larger number of markers (tens of thousands or higher) can be screened within a shorter time in comparison with the use of SSR markers . The higB. napus . A majorB. napus . In addiB. napus –30.B. napus [B. napus inbred lines, of which 73 were swedes or rutabagas. Similarly, Diers and Osborn [B. napus lines including two rutabagas. Mailer et al. [B. napus. Zhou et al. [Brassica accessions into spring, semi-winter and winter ecotypes. Gazave et al. [B. napus accessions with 30,881 high quality SNP markers and reported three major subpopulations, of which the highest variance was found in the spring and winter samples. Whole genome sequencing has indicated that winter oilseeds, which include rutabagas, may be the original form of B. napus and that this crop may have multiple origins [Previous molecular studies indicated that spring oilseed rape, winter oilseed rape, fodder and vegetable types, and rutabagas formed separate clusters of B. napus –33. Bus B. napus  used 89 d Osborn  used 43 u et al.  used 30,e et al.  genotype origins , 30.Brassica breeding programs around the world. The genetic diversity and variability that exist within and among rutabaga accessions and populations from the Nordic countries have not been examined. Understanding this diversity is important for conservation and broadening the use of this important resource. Therefore, the aim of the present study was to use high-throughput genotyping with Brassica SNP markers to estimate genetic diversity in rutabaga accessions from five Nordic countries .One hundred seventy-one rutabaga accessions are available  from the Nordic Genetic Resource Center, Alnarp, Sweden. Of these, 145 accessions are from the Nordic countries, 20 are from France, four are from Germany and one accession each is from Estonia and the United Kingdom. Many of these are landraces with great genetic variability that can be exploited in rutabaga and other Brassica chip were used to screen the 124 rutabaga accessions and three rutabaga cultivars. Among these, 31% (4213 SNPs) were monomorphic, 5% (701 SNPs) were low coverage site markers, and 14% (1929 SNPs) were missing data points for > 5% of the accessions. Thus, filtering removed ≈ 50% of the SNP markers, while the remaining ≈ 50% (6861 SNPs) were retained for the diversity analysis. This comprised 4390 A-genome and 2471 C-genome SNP markers.Thirteen thousand seven hundred four SNP markers on the 15 K SNP Figure P) detected separately for the NOR-, SWE-, FIN- and DNK- subpopulations was significantly higher (range 88.5–99.6%) than for the ISL-subpopulation (67.9%) (P < 0.05) (Table Na) was highest in the SWE-subpopulation (2.236 ± 0.005) and lowest in the ISL-subpopulation (1.707 ± 0.006) (Table Ne) and Shannon’s information index (I) were significantly higher in the SWE-subpopulation  compared with the ISL-subpopulation  (Table Na Freq ≥ 5%) and mean number of common alleles found in ≤50% of the subpopulations (Na common ≤ 50%) were lowest for the ISL-subpopulation .The proportion of polymorphic loci (%5) Table . The mea6) Table . Similary) Table . In addiThe diversity of the SNP markers expressed as the polymorphic information content (PIC) is presented in Fig. The number of SNP markers with minor allele frequency (MAF) ≤ 0.1 was of the order ISL- (4106 ≈ 60%) > FIN- 2115 ≈ 31%) > DNK- (1690 ≈ 25%) > NOR- (1518 ≈ 22%) > SWE-subpopulations (933 ≈ 14%). Thus, the frequency of minor alleles was highest for the ISL-subpopulation, intermediate for the FIN-, DEN- and NOR-subpopulations, and lowest for the SWE-subpopulation , also called gene diversity (D), followed similar patterns as the rest of the parameters measured with the exception of the MAF  of the rutabaga accessions from each country suggested that there was no significant difference in the genetic variability of the rutabaga accessions from Sweden (0.345 ± 0.002), Denmark (0.301 ± 0.002), Norway (0.292 ± 0.002), and Finland (0.288 ± 0.002). These accessions were, however, genetically different from the accessions from Iceland (0.191 ± 0.002)  Table .FST) are presented in Table FST values for all 10 pairwise combinations of all five subpopulations ranged from 0.032 to 0.133. Pairwise FST values for NOR/SWE, NOR/FIN and SWE/FIN ranged from 0.032 to 0.067 (lowest); the values for NOR/DNK, SWE/DNK and FIN/DNK ranged from 0.050 to 0.88 (intermediate); whereas the FST values for the ISL/NOR, ISL/SWE, ISL/DNK and ISL/FIN ranged from 0.103 to 0.133 (highest). Overall, the lowest FST value was found between the SWE- and FIN-subpopulations and the highest between the ISL- and DNK-subpopulations (Table Pairwise comparisons of population differentiation using the fixation statistics index (P = 0.108)  of the distance matrices obtained with TASSEL  and GenAlEx software for the rutabaga accessions were highly correlated Tables a and S2b08) Fig. A. This sPT) between the populations, however, revealed a higher genetic variance (18 to 27%) between the ISL-subpopulation and the NOR-, SWE-, FIN- and DNK-subpopulations (Table PT = 27%), followed by accessions from Iceland and Finland (ΦPT = 24%). In contrast, rutabaga accessions from Sweden and Finland were the most similar (ΦPT = 2%) followed by accessions from Norway and Sweden (ΦPT = 7%).Pairwise comparison of the AMOVA  based on the 6861 SNP markers clustered the 124 rutabaga accessions into six heterogeneous subgroups Fig. B using tThe neighbour-joining (NJ) based on the 6861 SNP markers clustered the 124 rutabaga accessions into four major branches Fig. C. The unThe unweighted pair group method with arithmetic mean (UPGMA) based on the 6861 SNP markers indicated that the trees for the 124 rutabaga accessions were clustered into five major branches Fig. D. The acOverall, the three multivariate analyses (PCoA + NJ + UPGMA) suggested the existence of four to six groups in the rutabaga accessions. However, correlations with their geographic origin were very low, except for the accessions from Iceland.The unrooted trees used to depict the NJ and UPGMA do not imply a known ancestral root of the three out-groups  determined following the method of Evanno et al. [∆K statistic values of K = 2 to 9, while the four alternative statistics  determined following Puechmaille [∆K statistic values, there was no significant difference in STRUCTURE run # 1, 2, 3 and 8 for analysis done with the accessions unassigned and for analysis with the accessions assigned to their respective countries of origin. In contrast, significant differences were found for STRUCTURE run # 4, 5, 6, 9, 10 and 11. The two methods produced approximately the same number of clusters (K = 3 to 4) at Burn-in and MCMC lengths each of 50,000 and at K = 1–10 and for 10 replicates (i.e. run #7)  against the number of clusters suggested the presence of subpopulations in the accessions  in the same cluster as ‘Wilhemsburger’ from Germany, which was used as an out-group   in the same cluster  Fig. D. In theter Fig. D. TherefA comprehensive body of literature exists on rutabagas in the main Nordic languages . This probably reflects the transmission of seeds and information on agronomic practices for rutabaga cultivation in the Nordic region since medieval times . TuressoB. napus, whereas Mailer et al. [In this study, SNP markers and combinations of allele- and distance-based population genetics statistics, multivariate clustering and Bayesian methods were used to examine genetic diversity and differentiation in rutabaga accessions from Norway, Sweden, Finland and Denmark and Iceland. Diers and Osborn  used rutr et al.  and Bus r et al.  comparedr et al.  and Zhour et al.  identifir et al.  and Lu er et al.  obtainedThe average of 2.012 ± 0.003 alleles per SNP locus obtained in this study was less than the 4.78 alleles per SSR locus found by Bus et al. . This was et al. , likely s et al. , barley s et al. , mushroos et al.  and othes et al.  reportedFST) obtained in the current study ranged from 0.032 to 0.133, which was within the 0.054 reported by Bus et al. [Brassica species [FST values for ISL-DNK (0.133), ISL-FIN (0.124), ISL-SWE (0.106) and ISL-NOR (0.103) compared with the pairwise FST values for DNK, FIN, NOR and SWE (range 0.032 to 0.088) may reflect enrichment caused by lack of mating between the Icelandic sub-population and the rest of the subpopulations. As an island in the North Atlantic, Iceland is geographically isolated from the other Nordic countries. This isolation, combined with possible differences in microclimatic and soil conditions, may have resulted in less exchange of germplasm between Iceland and Norway, Denmark, Finland and Sweden.The pairwise fixation index  of admixtures detected by the Bayesian population structure analyses. Inspection of clusters obtained by the use of the four genetic structure analysis methods showed that the majority (90–100%) of accessions with similar names were in the same group irrespective of their countries of origin. This suggests that the 6861 SNP markers were able to reliably determine the number of clusters in the 124 rutabaga accessions from the Nordic countries. The population structure analyses were in agreement with the allele diversity summary statistics obtained in this study.The hierarchical clustering of the genotype data by the use of NJ, UPGMA and PCoA analyses yielded 4 to 6 subgroups in the rutabaga accessions. Previous studies , 47 repo MaxMeaK  to the ∆ MaxMeaK  to decidB. napus.Our results showed that the majority of the genetic differences in rutabaga accessions from the Nordic countries were present within the Icelandic subpopulations, while accessions from Norway, Sweden, Finland and Denmark were genetically very similar. Given these findings, based on molecular genetics analyses, there may be value in additional and more detailed study of the morphological traits of accessions originating from these different countries. Lastly, these rutabaga accessions should be of great interest to breeders for increasing genetic diversity in One-hundred twenty-four rutabaga accessions collected from the Nordic Genetic Resource Center and used for genome wide association studies by Fredua-Agyeman et al. 12) were included in this study. These consisted of 23 accessions from Denmark, 12 from Finland, 11 from Iceland, 28 from Norway and 50 accessions from Sweden  < 0.05, and SNPs with missing data for > 5% of the accessions. Six thousand eight hundred sixty-one SNP markers were retained for the calculation of the genetic diversity indices and the population structure analyses. These comprised 4390 A-genome and 2471 C-genome SNP markers distributed across all 19 chromosomes of B. napus .P), the mean number of alleles per locus (Na), the mean number of effective alleles per locus (Ne), the mean expected heterozygosity (e), the mean unbiased expected heterozygosity (Ue), the mean number of alleles with a frequency ≥ 5% (Na Freq ≥ 5%), mean number of common alleles found in ≤25% and ≤ 50% of the subpopulations  and Shannon’s information index (I), within and among subpopulations, as well as Wright’s [FST) between the populations were determined with GenAlEx 6.5 [FST values were assessed at 1000 random permutations across the 6861 loci.The proportion of polymorphic loci  also called gene diversity (D) [In addition, the polymorphism information content (PIC), minor allele frequency (MAF) and the expected heterozygosity at any given locus (sity (D)  were evasity (D) .The genetic and similarity distance matrices within and among the subpopulations were calculated for the 6861 SNP markers and the 124 accessions using both GenAlEx 6.5 , 52 and The matrices were used to test the hierarchical partitioning of the analysis of molecular variance (AMOVA) among regions, populations and within accessions and their level of statistical significance was assessed based on 10,000 permutations . In addiSTRUCTURE v2.3.4 [K) based on the 6861 SNP markers. In addition, STRUCTURE was used to assign the rutabagas ‘Laurentian’ (Canada), ‘Wilhemsburger’ (Germany) and ‘Krasnoselskaya’ (Russia) into the Nordic subpopulations showing similar variation patterns.A Bayesian clustering approach, applying a Markov Chain Monte Carlo (MCMC) algorithm implemented in the population-genetic software E v2.3.4 , was useSTRUCTURE was run using the admixture model with correlated allele frequencies at a series of burn-in lengths from 5000 to 100,000 iterations and MCMC run lengths from 5000 to 100,000 permutations. In addition, STRUCTURE was run with the accessions unassigned to any population or country and with the accessions assigned to their countries of origin. Runs for each K = 1–10 were replicated 10 or 20 times. These were done to determine the parameters needed to reach convergence. The most likely number of clusters (the ad hoc ∆K test) and average log-likelihood plots were determined following Evanno et al. [K were used to group subpopulations into clusters with STRUCTURESELECTOR [o et al.  with STRo et al. . SecondlSELECTOR , 36. AccP ≤ 0.05) using SAS v. 9.4 .Statistical significance between means of the parameters  was established by Fisher’s protected least significant difference (LSD) test (Additional file 1.


Retractions are a key proxy for recognizing errors in research and publication and for reconciling misconduct in the scientific literature. The underlying factors associated with retractions can provide insight and guide policy for journal editors and authors within a discipline. The goal of this study was to systematically review and analyze retracted articles in veterinary medicine and animal health. A database search for retractions of articles with a veterinary/animal health topic, in a veterinary journal, or by veterinary institution-affiliated authors was conducted from first available records through February 2019 in MEDLINE/PubMed, Web of Science, Scopus, Retraction Watch, and Google Scholar. Annual frequency of retractions, journal and article characteristics, author affiliation and country, reasons for retraction, and retraction outcomes were recorded.P = .0155). Authors/groups with ≥2 retractions accounted for 37.2% of retractions. Authors from Iran and China published 19.4 and 18.2% of retracted articles respectively. Authors were affiliated with a faculty of veterinary medicine in 59.1% of retracted articles. Of 242 retractions, 204 (84.3%) were research articles, of which 6.4% were veterinary clinical research. Publication misconduct  accounted for 75.6% of retractions, compared with errors (20.6%) and research misconduct (18.2%). Journals published by societies/institutions were less likely than those from commercial publishers to indicate a reason for retraction. Thirty-one percent of HTML articles and 14% of PDFs were available online but not marked as retracted.Two-hundred-forty-two articles retracted between 1993 and 2019 were included in the study. Over this period, the estimated rate of retraction increased from 0.03/1000 to 1.07/1000 veterinary articles. Median time from publication to retraction was 478 days (range 0-3653 days). Retracted articles were published in 30 (12.3%) veterinary journals and 132 (81.5%) nonveterinary journals. Veterinary journals had disproportionately more retractions than nonveterinary journals (The rate of retraction in the field of veterinary and animal health has increased by ~ 10-fold per 1000 articles since 1993, resulting primarily from increased publication misconduct, often by repeat offenders. Veterinary journals and society/institutional journals could benefit from improvement in the quality of retraction notices.The online version contains supplementary material available at 10.1186/s12917-022-03167-x. Publication and research misconduct are important ethical concerns affecting the integrity of the biomedical literature –3. RetraA 2007 survey of science journal editors found that most editors were relatively unconcerned about publication ethics . HoweverA focus on retractions by discipline draws attention to sources of error and ethical misconduct relevant to a specific research community and can guide discipline-specific education and mentoring in scientific research and publishing. Investigations into retractions have been reported for a wide range of biomedical disciplines and medical specialties, including human-subject research , cancer The goal of this study was to comprehensively review and analyze retracted articles in the field of veterinary medicine and animal health. The frequency of retractions, journal and article characteristics , author affiliations, reasons for retraction, and retraction outcomes and trends over time were determined. Additional goals were to compare publisher type, open-access journals, veterinary journals, and veterinary faculty-affiliated authors with other journals and authors. The results of this study will identify sources and trends in errors and in ethical misconduct (research and publication misconduct) and how they are reconciled in the veterinary and animal health literature, providing insight to publishers, journal editors, and authors, including veterinarians and animal health scientists.n = 188, 77.6%), were published in a veterinary journal , and/or at least one author was affiliated with a veterinary faculty or a veterinary laboratory, department, or institute . Two-hundred-twenty of 242 (90.9%) retracted articles were indexed in MEDLINE or were in PubMed Central, including 141 from the initial PubMed search and 78 retractions identified initially through other databases. Retractions not initially identified in PubMed occurred prior to 2013 when affiliations were included only for the first author; 12 retracted articles from other databases that were found in PubMed were not identified as retracted.Of 548 citations retrieved, 306 were excluded and 242 were included in the study Fig. . The 242Of the 242 database citations for retracted articles, 201 were labeled as retracted, 1 was labeled as partially retracted, and 40 were labeled as withdrawn. Withdrawn articles are generally defined as articles retracted while in press or after early online publication, but use of the term was sometimes inconsistent: eight articles identified as retracted in the database were identified at the journal or article level as withdrawn; and 10 articles identified as withdrawn were identified at the journal or article level as retracted. Year of retraction was based on the electronic retraction date (when available) or the print retraction date; a date of retraction was not found for 14 articles. Six of the 14 articles (42.8%) without a date of retraction were in unindexed journals found only in Google . The number of retractions per year from 1993 through 2018 ranged from 0 to 48 , accounting for 23.5% of retractions but only 12.3% of journals .The 242 retractions were published in 162 unique journals published by commercial publishers, self-published by societies/institutions, or society journals hosted on a publisher website (Table P = .0258).Fifteen of 162 (9.2%) journals had ≥3 retractions each, accounting for 31.0% (75/242) of retracted articles (Table P < .0001). No correlation or difference in impact factor was found among journals based on the number of retractions.One-hundred thirty-three of 162 (82.1%) journals had a 2017 impact factor, with a median of 2.476 (range 0.217–41.058). The median impact factor of veterinary journals  was significantly lower than for nonveterinary journals  (P < .03). Fifty of 242 (20.6%) retractions had authors affiliated with a veterinary laboratory, department, or institute, of which 26 (52.0%) were first authors.A total of 143/242 (59.1%) retractions had faculty of veterinary medicine (FVM)-affiliated authors, of which 107/242 (44.2%) were first authors . The country affiliations of co-authors were similar to those of first authors, with 42 unique countries and 9 of the same 10 countries having the most retractions.First authors were from 37 countries. Iran and China accounted for the most retractions, and together with other top 10 countries accounted for 78.1% (189/242) of retracted articles (Table P < .0001). Retractions by repeat offenders accounted for a higher proportion of retractions due to publication misconduct  and a lower proportion of retractions due to errors  (P < .0001). A higher proportion of retractions by repeat offenders were published in nonveterinary  vs veterinary journals  (P = .0083).Authors or author groups with 2 or more retractions (repeat offenders) accounted for 37.2% 90/242) of retracted articles , free-ranging wildlife , client-owned pets , private or commercial farm animals , institutional herds or colonies , aquacultured fish , abbatoirs , and zoos . Twelve (6.6%) articles did not state the source of animals used. Six (3.3%) articles used human subjects or donors.Most (84.3%) retractions were research articles, with fewer case reports/case series and review articles Table . Nearly n = 228). Commercial publisher-hosted society journals had a significantly longer interval to retraction (median 789.5 days) compared with commercial publishers (243.5 days) and society/institutional publishers (472.5 days) (P = .0296). No significant difference in time to retraction was found for open-access or veterinary journals. Median days to retraction was significantly shorter for articles labeled withdrawn  as compared with those labeled retracted  (P < .0001). No correlation was found between the time to retraction and journal impact factor.The median time interval between publication and retraction was 477.5 days . A significantly higher proportion of veterinary journals did not state the reason for retraction  compared with nonveterinary journals  (P < .0001). A higher proportion of journals published by societies/institutions did not state the reason for retraction  compared with commercial publishers  (P = .0013). No difference was found in the reason for retraction between open-access and non-open-access journals. Median journal impact factor was significantly higher for articles retracted for research misconduct (4.011) compared with those retracted for errors (2.525) or publication misconduct (2.275) (P < .0001).Reasons were provided for retraction in 207/242 (85.5%) articles Table . The mosn = 11), government agencies (n = 2), and industry (n = 1) were conducted primarily in cases involving research misconduct; formal internal investigations by journals or publishers  were conducted, primarily for cases of publication misconduct . Similarly, a higher proportion of society/institutional-published journals  vs commercial publishers  did not specify who initiated retraction (P = .0021). Higher proportions of retractions were initiated by editors/publishers (vs authors) in open-access journals (58.0% vs 34.5%) and in cases of publication misconduct (62.6% vs 32.1%) (P < .01).The party initiating retraction was indicated in 227/242 (93.8%) of retraction notices; of these 14 indicated “author and/or editor” so the initiating party was not clear (Table P = .0009). Forty-five percent (110/242) of HTML articles lacked any indication of retraction or had been removed from the website; 31.4% (101/242) of PDFs lacked any indication of retraction or had been removed from the journal website. A higher proportion of open-access journals  had unmarked article PDFs on the journal website compared with non-open-access journals  (P = .0234).A majority of retracted articles  were accompanied by either a descriptive retraction notice or a link to a retraction notice. Society/institutional publishers were more likely than commercial publishers to simply state ‘retracted’ in the article title rather than provide a descriptive retraction notice  for comparison of trends over time. The percentage of HTML and PDF articles marked as retracted increased significantly in 2016-2018 compared with prior years (P < .01) . Significantly more retractions were the result of research misconduct (vs publication misconduct) in 2012-2015 (P = .0004). Median time from publication to retraction did not differ significantly in 1993-2011 (302 days) compared to 2012-2015 (409 days) and 2016-18 (686 days) (P = .1667).Retraction outcome data were examined by year group  accounting for more than 75% of retracted articles. Notably, a few repeat offenders accounted for a high proportion of total retractions. In part because of this, veterinary journals had significantly more retractions than nonveterinary journals, and together with society/institutional journals, were less likely to provide informative retraction notices. These findings suggest that enhanced editorial oversight of peer review, and timely and transparent recognition of ethical misconduct, could help correct the literature and minimize the societal impact of retractions on the discipline.Veterinary medicine and animal health is a diverse field that can be challenging to capture in bibliographic analyses. The comprehensive approach used herein included not only the veterinary journal category (across three major indexers for complete coverage), but also retractions with a topical focus on animal health and disease and those with authors at veterinary institutions, many of whom publish in non-veterinary journals.  Unlike retractions included in the Veterinary Science category of Retraction Watch , the preAs reported in other bibliographic analyses –8, the rRetractions in veterinary medicine and animal health were most frequently the result of publication misconduct, which has been associated with reduced barriers to publication  and inadequate editorial oversight. The proportion of retractions due to plagiarism and duplicate publication combined 44%) was similar to that reported in other studies and disciplines 42-48%) , 19, 22.4% was si% , 19, 2Veterinary journals comprised surprisingly few of the journals in this study, perhaps because of the predominance of basic  science articles retracted and because of the lower impact factors or perceived value of veterinary journals compared with other biomedical journals , 28. AltAuthor conduct and publication decisions are influenced by training, mentorship, affiliation, institutional culture, economic incentives, and national and academic ethical policies , 23, 31.Diagnostic Pathology [Reproduction in Domestic Animals, a veterinary journal appropriate for the research topic. The proliferation of low-quality, open-access journals can provide opportunity for repeat offenders to exploit new and unsupervised systems [Repeat offenders accounted for nearly 40% of retractions in veterinary medicine and animal health. Indeed, those with ≥5 retractions comprised nearly 25% of total retractions, more than double the 10% expected based on a power-law model of repeating probability . This hiathology , 22. In athology . However systems , but higRetractions reflect a failure to identify problems prior to publication but also signal a willingness on the part of journals and authors to correct mistakes . RetractIt is recommended that a retraction notice also state the parties who initiated, issued, and supported the decision to retract; whether the authors were contacted and agreed to the retraction; the reason(s) for retraction; the sections retracted and the effect on the rest of the article; and the action taken by the journal . Few retWhile the scholarly impact of articles and authors decreases after retraction, the societal impact of retractions on the field of veterinary science and public health can be substantive , 41. ConRetractions of journal articles in the veterinary medicine and animal health literature have increased substantially since 2012, primarily as a result of publication misconduct. Retracted articles reflect the breadth of animal health research and its diverse investigators, including both FVM and nonveterinary affiliations. Repeat offenders have a disproportionate effect on retractions, but recent trends suggest increasing editorial assertion and implementation of policies that better identify retracted literature. The societal impact of retractions on the animal health profession can be minimized through rigorous peer review processes, enhanced editorial oversight, use of digital tools, and the timely identification and transparent description of the error or misconduct.A systematic search for retracted articles was conducted in February 2019 using five databases: PubMed , Web of Science (Science Citation Index Expanded), Scopus, Retraction Watch, and Google Scholar. In PubMed, the Web of Science, and Scopus, a search on the term “veterinary”  was filtered by document type “retracted publications” and “retraction of publication”  for the entire time period of available articles (through February 2019). In Retraction Watch, retracted articles in the topic field “(HSC} Veterinary Science” were retrieved . Google Retracted articles in English were included if they met at least one of the following inclusion criteria: (1) the article was published in a journal indexed as a “veterinary” journal by MEDLINE, Science Citation Index, or Scopus; (2) at least one of the authors was affiliated with a faculty (school/college) of veterinary medicine (FVM), or with a veterinary laboratory, department, or institute; and (3) the article involved veterinary medicine or animal health/disease . When retraction notices listed additional articles by the same author(s) that had not already been retrieved in the search process, the additional articles were examined and added to the study if they met one or more of the inclusion criteria. Withdrawn articles, defined as articles retracted while in press or after early online publication, were included because of inconsistent and sometimes interchangeable use of the terms ‘retraction’ and ‘withdrawal’ by journals and indexers . ArticleArticles were given a unique identifier and the original citation and source (index) were recorded. PubMed was used as the core source; retraction records found initially in Web of Science, Scopus, Retraction Watch, and Google were subsequently also searched for in PubMed. Because more than 90% of retracted articles ultimately were found in PubMed, the PubMed dataset of “veterinary” articles was used to estimate the prevalence of retractions. Retracted articles, journal websites, and retraction notices were retrieved and examined. Data were recorded for variables associated with the journal, the article, and the retraction notice Table . JournalP value of <.05 was considered as significant.Data were compiled in an Excel spreadsheet  and analyzed statistically using JMP software . Based on visual assessment and the Anderson-Darling goodness-of-fit test, distribution of continuous variables  was found to be non-Gaussian. Therefore, those quantitative results were expressed as median and range  and the nonparametric Wilcoxon rank sum test was used to compare results between groups. Categorical variables were expressed as frequency (%), and proportions were compared using Chi-square analysis. A Additional file 1. Retracted articles included in this study.


Postoperative diarrhea, including high-output stoma (HOS), frequently occurs after colorectal surgery; its risk factors and clinical implications on subsequent complications remain unknown. This study aimed to evaluate the risk factors and clinical implications of postoperative diarrhea after primary colorectal cancer (CRC) surgery.This prospective observational study included patients with CRC who underwent radical surgery at six hospitals between June 2016 and December 2017. The patients were categorized into three groups .P = 0.012 and P = 0.0274, respectively). Postoperative diarrhea was more likely to occur in patients with rectal cancer than in those with colon cancer (P = 0.0501). In the non-stoma and ileostoma groups, the probability of the occurrence of other complications with Clavien–Dindo (C-D) grades II or higher was significantly higher in patients with C-D grade I diarrhea, including HOS, than in patients without diarrhea .A total of 178 patients participated in the study. In the non-stoma group, the incidence of postoperative diarrhea was 18.4% (27/147). The incidence of HOS was 28.6% (4/14) in the ileostoma group, and 0% in the colostoma group. Multivariable analyses of the incidence of diarrhea in the non-stoma group indicated that habitual smoking and hypertension were significantly associated with postoperative diarrhea (Smoking and hypertension are the independent predictors of postoperative diarrhea after an elective CRC surgery. Rectal cancer surgery seems to be associated with postoperative diarrhea more than colon cancer surgery does. Mild postoperative diarrhea may lead to more severe complications. Clostridioides difficile (formerly known as Clostridium difficile) infection (CDI) is also a common nosocomial enteritis that causes diarrhea and occasionally severe symptoms [Postoperative diarrhea is a frequent complication that can negatively affect the nutritional status and clinical outcome of patients after colorectal cancer (CRC) surgery. Various causes of postoperative diarrhea include bowel stenosis due to a huge tumor, surgical procedures, and perioperative pharmacological treatment , 2. Clossymptoms , 4. Meansymptoms , 6. Accosymptoms –7.Over the last decade, many retrospective studies have been conducted to elucidate the chronic gastrointestinal complications after a CRC surgery, including diarrhea , 8, 9. HThis  is a multicenter, prospective, observational study involving patients with postoperative diarrhea who underwent CRC surgery at six hospitals in Japan. A total of 178 consecutive patients with CRC, who were scheduled for an elective radical surgery between June 2016 and December 2017, were enrolled in this study. The study was performed in accordance with the ethical standards as laid down in the 1964 Declaration of Helsinki, and was approved by the Research Ethical Committee of Shiga University of Medical Science (approved IRB number: 27–222). Written informed consent was obtained from all eligible patients.The inclusion criteria were as follows: (1) diagnosis of histopathologically confirmed CRC, (2) scheduled for curative resection, (3) age ≥ 20 years at the time of enrollment, and (4) performance status (PS) ≤ 2 on the Eastern Cooperative Oncology Group (ECOG) scale. Those with diarrheal symptoms before surgery, a history of CDI, or an emergency operation were excluded. Patients with severe complications or a PS ≥ 3 on the ECOG scale were also excluded, due to the likelihood of fecal incontinence. Bowel preparation before the surgery was performed at the discretion of each facility. This trial was registered with the UMIN Clinical Trials Registry . This manuscript has been prepared in accordance with the STROBE guidelines .Clostridioides difficile (CD) toxin assay and fecal culture examinations were performed at the discretion of the attending physician. Other adverse events and postoperative complications, including surgical site infection (SSI), were also assessed according to the Clavien–Dindo (C-D) classification system [The patients were categorized into three groups according to stoma creation: 1) non-stoma group, 2) colostoma group, and 3) ileostoma group. The definition of postoperative diarrhea for the non-stoma group in this study was adopted on the basis of the World Health Organization’s criteria; accordingly, postoperative diarrhea was defined as a “frequent watery bowel movement for more than three times per day within postoperative 30 days ”. Diarrhn system .We analyzed the patients’ background characteristics  and preoperative care factors . Data on the postoperative outcomes (including the onset date and treatment of postoperative complications) and postoperative hospital days were prospectively collected. Analyses of risk factors for postoperative diarrhea were performed for the non-stoma group.Furthermore, the clinical impact of mild postoperative diarrhea on other complications that followed was evaluated. The associations between C-D grade I diarrhea, including HOS, and other complications of C-D grades II and higher were investigated in the non-stoma and ileostoma groups; in both groups, the patients received bowel reconstruction. In this investigation, the patients with diarrhea of C-D grade II or higher were excluded. Furthermore, the patients who had diarrhea after other C-D grade II or higher complications were excluded in consideration of therapeutic antibiotic-induced diarrhea.P < 0.05) in the univariate analysis; factors predictive of postoperative diarrhea were determined. The analysis was performed using the backward stepwise regression method. A P-value of < 0.05 was considered to indicate statistical significance. Statistical analyses were carried out using the statistical program EZR , which is a graphical user interface for R  [The categorical variables are expressed as numbers and percentages. In the univariate analysis, group comparisons were performed using the Fisher’s exact test and the chi-squared test for binary variables. Multivariable logistic regression analysis was performed using variables found to be significant (Austria) .2; 72 patients (40.4%) were smokers, 69 patients (38.8%) had a high blood pressure, and 28 (15.7%) had diabetes mellitus. Preoperative medications were as follows: therapeutic antibiotics , antacid , and chemotherapy . No patients underwent perioperative radiation therapy. The median preoperative hospital stay duration was 3 days. Before the surgery, 157 patients (88.2%) underwent mechanical bowel preparation, while 61 patients (34.3%) underwent chemical bowel preparation.A total of 178 patients were enrolled in this study; of these, 147 were categorized into the non-stoma group. Based on the stoma-creation surgery, 17 and 14 patients were categorized into the colostoma and ileostoma groups, respectively. Table n = 60), left colon (n = 42), rectum (n = 75), and both the right colon and the rectum (n = 1). The TNM classification stages were as follows: 0 (n = 1), I (n = 45), II (n = 32), III (n = 74), and IV (n = 26). Among the 178 patients, 127 (71.3%) underwent laparoscopic surgery. The median operation time and blood loss were 240 min and 50 mL, respectively. Eleven patients (6.2%) received blood transfusion during the operation.The tumor locations were as follows: right colon . Among the stoma groups, HOS occurred with an incidence of 28.6% (4/14) in the ileostoma group, while it did not occur in the colostoma group. Out of the 31 patients with postoperative diarrhea including HOS, 17 patients (54.8%) underwent the rapid CD toxin assay and fecal culture examination. All the patients tested negative for the CD toxin. However, two patients (11.8%) demonstrated CD-antigen positivity; one of these also had a positive fecal culture on examination. Both patients were clinically diagnosed with mild CDI (one on the third postoperative day and the other on the fourth postoperative day). They received metronidazole, in accordance with the clinical practice guidelines .Twenty-nine patients with diarrhea due to causes other than CDI were all mildly ill and did not require special treatments, such as antibiotics; hence they were judged to have a C-D grade I postoperative complication. The median onset of diarrhea was postoperative day 4 . The median duration of postoperative hospitalization was 13 days in both patients who had diarrhea  and those who did not . In this study, patients with diarrhea did not take any postoperative aperients before the onset of diarrhea.n = 147), univariate analysis of various clinical factors for postoperative diarrhea was performed In the non-stoma group (ed Table . In the P = 0.012) and hypertension  were the independent predictive factors for postoperative diarrhea (Table P = 0.0501).The multivariable analysis included the three variables that showed statistical significance in the univariate analyses, i.e., smoking habits, hypertension, and tumor location (rectum). Among these, smoking habits .Finally, we investigated the impact of C-D grade I diarrhea on other complications of C-D grade II or higher. During the study period, no C-D grade IV life-threatening complications or C-D grade V complications (such as surgery-related deaths) were noted. There were no cases where C-D grade II and III complications occurred simultaneously. In this investigation, a patient with antibiotic-induced postoperative diarrhea after incisional SSI was excluded. Furthermore, two patients with C-D grade II CDI were excluded. The total incidence of C-D grade II and III complications in patients with C-D grade I diarrhea was 39.3% (11/28); this was significantly higher than that in patients without postoperative diarrhea , including Crohn’s disease, collagenous colitis, and lymphocytic colitis , 17. BirRecently, some retrospective studies have reported that postoperative diarrhea impacts postoperative outcomes after colorectal surgery , 22. GaeOur data suggest that the patients who undergo rectal cancer surgery tend to experience postoperative diarrhea more commonly as compared to patients who undergo colon cancer surgery. However, this finding contradicts the findings of previous studies; right hemicolectomy is recognized to be associated with chronic postoperative diarrhea due to residual bile, which acts as an osmotic agent, and disruption of the ileal brake , 8, 24. Perioperative oral synbiotic supplementation is expected to improve the digestive microbial balance and present postoperative infectious complications. Despite various reports on probiotics, prebiotics, and synbiotics, their efficacies in preventing CRC surgery complications remain controversial . KomatsuThere is no agreed definition of HOS; it refers to excretion ranging from 1500 to more than 2000 mL per day or for 2 consecutive days. The non-standardized definition in this study may have influenced the results. Moreover, this prospective study has several limitations, including the small number of patients, different operative procedures, absence of standardized bowel preparation, and difficulty in diagnosing CDI (due to the low sensitivity of the CD toxin assay). Finally, this study lacks a sufficient number of cases, which reduces the generalizability of its results. The courses and mechanisms of postoperative diarrhea are diverse and complex; therefore, further basic physiological and pathological studies may be needed along with clinical studies to identify ways to prevent postoperative diarrhea.Our findings suggest that smoking and hypertension are independent predictors of acute postoperative diarrhea, and mild postoperative diarrhea may lead to other more severe complications (including UTI) after an elective CRC surgery, especially rectal cancer surgery. Additional studies with much larger participant numbers are necessary to confirm the relationship between postoperative diarrhea and other complications in CRC surgery.


In total, 2.2% of patients developed contralateral breast and 1.6% other carcinomas. Locoregional recurrence (LR) occurred in 3.1% of patients (0% for in situ carcinomas). Positive margins after surgery, extracapsular node invasion, and HER2-enriched/triple-negative tumors were linked with significantly worse LR-free survival. The involvement of more than three nodes and luminal type other than A were independent prognostic variables of metastasis and death events. HypoAR delivering a biological dose of 50–52 Gy to the breast/ASA is a safe and effective therapy for patients treated with conservative surgery. The risk of carcinogenesis is low. Positive surgical margins, extracapsular node invasion, and HER2-enriched/triple-negative phenotypes appear as a cluster of features linked with a higher risk for locoregional relapse.We report long-term results (median follow-up 12 years) of hypofractionated accelerated radiotherapy (HypoAR) in patients treated with breast-conserving surgery. In total, 367 women were treated with HypoAR. Axillary and supraclavicular area (ASA) were treated in patients with involved nodes. In total, 290 patients (scheme A) received 3.5 Gy/day ×10 fractions (breast/ASA) followed by two 4 Gy fractions with electrons to the affected breast quadrant within 16 days. In total, 77 patients (Scheme B) received 2.7 Gy/day for 16 consecutive fractions (breast/ASA) within 22 days, while concurrently, the affected breast quadrant received an electron booster dose of 0.8 Gy for the first 13 fractions. Amifostine was offered to 252/367 patients. Early radiation toxicity was minimal. Regarding late toxicities, symptomatic breast edema was noted in 2.2%, asymptomatic breast fibrosis in 1.9%, and arm lymphedema in 3.7% of patients. Amifostine reduced early radiation dermatitis ( Conservation of the breast with partial mastectomy techniques followed by postoperative radiotherapy has been an established organ-preserving therapy during the last three decades . RadiothCondensing the radiotherapy schedule using hypofractionation has long been considered prohibited, after the ‘lessons from complications’ paper published by Fletcher GH in 1991, one of the founders of modern radiotherapy . DevelopOur early experience, published in 2002, treating breast cancer patients with a 3.5 Gy per fraction for 10 consecutive fractions followed by booster radiotherapy dose provided encouraging results  or modifFrom January 2003 to December 2010, 367 women with breast cancer treated with breast-conservative surgery were recruited in a prospective trial of hypofractionated and accelerated radiotherapy (HypoAR), focusing on tolerance and efficacy. The study was approved by the Ethics and Research Committee of the University Hospital of Alexandoupolis . Preliminary results of the trial were published in 2009 . The curAll patients had a performance status of 0 . Patients previously treated with radiotherapy in the chest area, pregnant women, patients with concurrent hematological or other malignancies, and patients with significant heart, lung, liver, renal, and psychiatric disease were excluded. All patients gave written informed consent. Radiotherapy was delivered with 3-D-conformal techniques using a 6–18 MV linear accelerator (ELEKTA), endowed with a multileaf collimator. Following CT simulation, treatment planning was performed with the ‘Plato’ (Nucletron) planning system. The breast was treated by applying two to four tangential X-ray fields. All patients with involved nodes received radiotherapy to the axillary and supraclavicular area with anteroposterior fields. None of the patients received internal mammary area irradiation. All patients received a boosted dose to the affected quadrant. Two different radiotherapy schemes were applied, according to the physician’s discretion, as follows:Scheme A (290 patients): Breast and axillary area (when included) received 3.5 Gy per day for 10 consecutive fractions within 12 days. Subsequently, the quadrant where the primary tumor was located received two additional fractions of 4 Gy with 10–15 MeV electrons (with appropriate adjustment with bolus according to the anatomy). The overall treatment time was 16 days.Scheme B (77 patients): Breast and axillary area (when included) received 2.7 Gy per day for 16 consecutive fractions within 22 days. The affected breast quadrant received a concomitant booster dose of 0.8 Gy (3.5 Gy/day to the quadrant) for the first 8 fractions. The overall treatment time was 22 days.Schedule B was formulated to simulate the Canadian schedule with concomitant booster dose to the affected quadrant. The choice between schedule A and B was at the discretion of the physician. Schedule B was more cumbersome, and it was soon abandoned.(T) = D [(α/β + d)/(α/β + 2)] + λ (Tc − To), where ‘Tc’ is the number of days required for the delivery of the EQD2 using a conventionally fractionated scheme, ‘To’ is the number of days required for the delivery of the current scheme, and ‘λ’ is the estimated daily dose consumed to compensate for rapid tumor repopulation. For cancer and normal breast area tissues, an α/β ratio of 4 Gy was considered as calculated by Yarnold et al. [The radiobiological dose analysis is reported in d et al. ,8. For cd et al.   Common Terminology Criteria for Adverse Events Version 5 scale was used to assess chemotherapy and acute radiation toxicity . The LENt-test for group analysis of continuous variables. Survival curves were plotted using the Kaplan–Meier method, and the log-rank test was used to determine statistical differences between life tables. A Cox proportional hazard model, including variables significant at univariate analysis, was used for multivariate analysis of locoregional relapse, metastasis, and death events. A p-value of < 0.05 was considered for significance.The GraphPad Prism 7.00 version package  was used to perform statistical analysis and graph presentation. The SPSS program was used to perform multivariate analysis. The Fisher’s exact test was used to compare categorical variables and the unpaired two-tailed p = 0.001) and breast edema (p = 0.03) but had no effect on the score of pain. Early toxicities were similar between patients receiving the schedule A and B. There was a trend for schedule B to produce higher rates of acute dermatitis (p = 0.06).Consistent with the preliminary reports , early tp = 0.09) and of breast edema (p = 0.06). There was no difference between schedule A and B.Late toxicities were also low . Of inteWithin up to 17 years of follow-up, 8 (2.2%) patients developed a new primary cancer of the contralateral breast. Six out of 367 (1.6%) patients developed non-breast carcinomas .Out of 12 patients with in situ carcinoma treated with HypoAR (6 patients with the regimen A and 6 with the regimen B), none of them presented with locoregional recurrence, and all patients were alive with no evidence of disease within 120–180 months of follow-up.Locoregional recurrence (LR) occurred in 11/355 (3.1%) patients , as shown in p = 0.02; p = 0.01; p = 0.26). Moreover, extensive lymphadenectomy (>16 dissected nodes) did not have any effect on LRFS (p = 0.97).The Kaplan–Meier locoregional recurrence-free survival (LRFS) curves and relevant univariate analysis are shown in p = 0.009; The group of patients with HER2-enriched or triple-negative tumors also had a higher risk of locoregional recurrence .At the time of analysis, 55/355 (15.5%) patients had been recorded with metastasis to distant organs . Kaplan–Meier univariate analysis showed that T1 stage, number of involved nodes less than four, and luminal type A were significantly related with better metastasis-free survival , as shown in At the time of analysis, 68/355 had died, 20/68 from reasons not related to their breast cancer and 48/68 from breast cancer progression. Kaplan–Meier univariate analysis of disease-specific events showed that T1 stage, number of involved nodes less than 4 (Nx patients excluded), and luminal type A were significantly related with better distant metastasis-free survival  regimen delivering 5.7 Gy/week . This reThe administration of a booster dose to the affected breast quadrant is a policy not always followed or, at best, adopted in patients with high-risk of local relapse, like patients with positive surgical margins, large tumors at diagnosis, high-grade tumors, or even tumors with adverse molecular features . This boAnother issue that remains to be resolved in the breast-HypoAR practice is the fractionation applied for the treatment of axillary and supraclavicular areas. The Standardisation of Breast Radiotherapy UK START A and B trials showed that arm edema and shoulder stiffness had no different fraction sensitivity than breast and chest . NeverthWe further analyzed the therapy features, and histological and molecular variables that affect prognosis. The two HypoAR schedules applied were equivalent in terms of local and metastasis-free survival rates and, moreover, amifostine did not have any effect on the efficacy of radiotherapy. Positive surgical margins and extracapsular nodal involvement were independent variables related to locoregional recurrence. Of interest, HER2-enriched and triple-negative tumors were related to increased locoregional relapse rates. These ominous prognostic features have also been identified in previous studies . The tumAnother issue that has been raised during the early era of breast cancer HypoAR studies is that long-term follow-up should also focus on an eventual higher carcinogenic risk of large radiotherapy fractions. Indeed, in an analysis by Kirova et al., radiotherapy for breast cancer has been associated with increased risk for the development of secondary carcinomas, especially of the lung, and sarcomas . AuthorsIt is concluded that HypoAR is a safe and effective therapy for patients treated with conservative surgery. A time-corrected EQD2 of 48–50 Gy proved optimal for both breast and axilla/supraclavicular (omission of sentence). The risk of carcinogenesis is at least as low as the one expected from conventional radiotherapy. Positive surgical margins, extracapsular node invasion, and HER2-enriched/triple-negative phenotype appear as a cluster of characteristics implying a high risk for locoregional relapse. Although the use of booster radiation to the affected breast quadrant is not a standard practice in many institutions, the current study provides evidence that this can be safely applied in subgroups of patients with high risk, but its therapeutic value needs evaluation in randomized trials.


As lung transplantation (LTX) is a valuable treatment procedure for end-stage pulmonary disease, delayed referral to a transplant center should be avoided. We aimed to conduct a single-center analysis of the survival time after listing for LTX and waitlist mortality in each disease category in a Japanese population.We included patients listed for LTX at Tohoku University Hospital from January 2007 to December 2020 who were followed up until March 2021. Pulmonary disease was categorized into the Obstructive, Vascular, Suppurative, Fibrosis, and Allogeneic groups. Risk factors for waitlist mortality were assessed using a Cox proportional hazards model. The Kaplan–Meier method was used to model time to death.p < .001), followed by the Allogeneic (35.0%), Suppurative (33.3%), Vascular (32.1%), and Obstructive (13.1%) groups. The Fibrosis group showed a remarkable risk for waitlist mortality .We included 269 LTX candidates. Of those, 100, 72, and 97 patients were transplanted, waiting, and dead, respectively. The median time to LTX and time to death were 796 days (interquartile range [IQR] 579–1056) and 323 days (IQR 129–528), respectively. The Fibrosis group showed the highest mortality (50.9%; In Japan, the waiting time is extremely long and candidates with Fibrosis have high mortality. There is a need to document outcomes based on the underlying disease for listed LTX candidates to help determine the optimal timing for listing patients based on the estimated local waiting time.The online version contains supplementary material available at 10.1186/s12890-021-01760-8. Advance in lung transplantation (LTX) has continued to evolve over the last decades and LTX has become a valuable option for selected patients with end-stage pulmonary disease. Delayed referral to a transplant center limits the opportunity for LTX; however, timely referral is difficult for most pulmonary diseases given the limited studies on listed individuals and limited evidence regarding proper consultation for LTX in each pulmonary disease , 2. ThisGiven the limited pool of donated organs, optimizing organ allocation to decrease waitlist mortality in patients listed for LTX is needed. The lung allocation score has decreased the mortality rate among individuals on the waitlist and increased the survival benefit for LTX recipients in the United States –5. HowevGiven the unique circumstances surrounding LTX in Japan, we aimed to conduct a single-center analysis of the survival time after listing for LTX and the waitlist mortality in each disease category. Moreover, we aimed to identify risk factors for mortality in LTX candidates on the waitlist.We included patients listed for LTX at Tohoku University Hospital (TUH) from January 1st, 2007 to December 31st, 2020, who were followed up to March 31st, 2021. Data were collected from application forms filled at the time of listing at TUH. Patients who refused LTX after listing or developed non-pulmonary irreversible complications were delisted from the registration file in JOTN. Outcomes  were reviewed on March 31st, 2021. Pulmonary disease was categorized into obstructive lung disease (Obstructive), pulmonary vascular disease (Vascular), suppurative lung disease (Suppurative), interstitial lung disease (Fibrosis), and allogeneic disease (Allogeneic). The Additional file p < 0.05. Statistical analyses and graph generation were performed using EZR  [Variables were presented as percentages or medians (interquartile range [IQR]), as appropriate. Among-group differences in categorical and continuous variables were compared using the chi-square test and Kruskal–Wallis test, respectively. Further, analyses based on IPF and non-IPF interstitial lung disease (ILD) were performed using the chi-square or Fisher’s exact test and the Mann Whitney test, respectively. Risk factors for waitlist mortality were assessed using a Cox proportional hazards model. Clinically important variables  and presumed risk factors for ILD  were selected for analysis. The Kaplan–Meier method was used to model time to death  where candidates who were waiting on March 31st, 2021 or transplanted in the study period were considered censored. Among-group differences were calculated using the log-rank test. Correlations between the predicted FVC and survival after listing were analyzed using Pearson’s correlation coefficient (r). Statistical significance was set at Between 2007 and 2020, 1507 patients were registered at JOTN for LTX in the whole of Japan ), followed by the Allogeneic (287 days [IQR 132–449]) and Fibrosis (315 days [IQR 113–513]) with no significant difference (p = 0.083).The pulmonary disease was categorized as Obstructive, Vascular, Suppurative, Fibrosis, and Allogeneic in 22.7% 61/269), 19.7% (53/269), 7.8% (21/269), 42.4% (114/269), and 7.4% (20/269) of the patients, respectively . Table Figure p = 0.039), male , and die (61.4% vs. 40.4% p = 0.039) in the study period (Table p = 0.314) and overall waiting time . Compared with non-IPF ILD, IPF was significantly associated with lower survival on the waitlist (Log-rank p = 0.41)  and non-IPF ILD (n = 57). Non-IPF ILD accounted for 50.9% (29/57), 10.5% (6/57), and 8.8% (5/57) of cases of CTD-ILD, drug-induced interstitial lung disease, and pleuroparenchymal fibroelastosis, respectively  has a poor prognosis; moreover, IPF has a worse prognosis even after listing for LTX –14. At oTo identify risk factors or outcomes in LTX candidates with similar pathophysiology, it would be useful to categorize pulmonary diseases into smaller groups. Pulmonary disease has been classified as A (obstructive disease), B (vascular disease), C (cystic fibrosis), and D (restrictive disease) in the United States, which has improved allocation equity and quality according to the Organ Procurement and Transplantation Network , 17. SinThis study has several limitations. First, this was a single-center retrospective study; therefore, our findings do not fully represent the current transplant circumstance in Japan. Some of the patients were listed a decade ago, with the LTX indications having changed over time. Currently, LAM is not often listed while IPF has become increasingly listed and is the most common indication for LTX. Therefore, there is a need for routine reports from the registry organization or academic societies regarding outcomes based on the pulmonary disease listed for LTX. Moreover, there is a need for a prospective study including all transplant centers in Japan to identify risk factors for waitlist mortality in each disease category or specific disease. The timing of referral or listing for LTX should be configured based on the probability of the mortality from the perspective of the waiting time in each region or country. Although our findings are too preliminary to conclude the optimal timing for listing patients with IPF or even the disease category, they could facilitate future multicenter trials.Our findings demonstrated that in Japan, the waiting time is exceptionally long and the mortality rate is substantially high among LTX candidates with Fibrosis. There is a need for routine reports documenting the outcomes based on the underlying disease in LTX candidates to help identify prognostic factors in patients on the waitlist, which could allow optimal timing for listing based on the estimated local waiting time. Although our findings are too preliminary to propose a timely listing for each pulmonary disease, our findings could facilitate future related studies.Additional file 1. List of pulmonary disease in transplant candidates in the study.