Fake news can impair memory leading to societal controversies such as COVID-19 vaccine efficacy. The pernicious influence of fake news is clear when ineffective corrections leave memories outdated. A key theoretical issue is whether people should recall fake news while reading corrections with contradictory details. The familiarity backfire view proposes that recalling fake news increases its familiarity, leading to interference. However, the integrative encoding view proposes that recalling fake news promotes co-activation and binding of contradictory details, leading to facilitation. Two experiments examined if one theory better accounts for memory updating after participants recalled actual fake news details when reading headlines that corrected misinformation. In Phase 1, participants read real and fake news headlines of unclear veracity taken from various internet sources. In Phase 2, participants read real news headlines that reaffirmed real news and corrected fake news from Phase 1. When they detected that Phase 2 real news corrected fake news, they attempted to recall Phase 1 fake news. In Phase 3, participants first recalled real news details. When they remembered that those details were corrections from Phase 2, they attempted to recall fake news from Phase 1. Recalling fake news when noticing corrections in Phase 2 led to better memory for real news in Phase 3 when fake news was recalled again and worse memory for real news in Phase 3 when fake news was not recalled again. Both views explain part of the memory differences associated with recalling fake news during corrections, but only when considering whether people recollected that fake news had been corrected.The online version contains supplementary material available at 10.1186/s41235-022-00434-1. The proliferation of fake news in the media can create inaccurate memories that lead to negative effects on public beliefs, decision making, and health. Correcting fake news to mitigate its effects is sometimes effective, but it can also lead to interference in memory when corrections promote the retrieval of misinformation. Corrections facilitate the replacement of outdated misinformation with accurate details , when people can recollect that those discrepancies had occurred. But corrections can also impede such updating when retrieving fake news details during corrections increases the feeling that misinformation details are familiar. The present study shows that these contradictory effects of corrections can be explained by a theory of memory updating proposing key roles for integrative encoding of real and fake news details, retrieval-enhanced familiarity of misinformation, and subsequent recollection of contradictory details and their relationship. The present findings suggest that successfully recalling fake news details when corrections are initially noticed can improve subsequent memory for real news. However, for this to be effective, recalled fake news must enable encoding processes that promote later recollection that fake news had been corrected. These findings support the recommendation that corrections should encourage comparisons of misinformation with accurate information, but only when the relationship between information is encoded well enough to be recollected.The problem of “fake news” is a present dilemma. Fake news is false information presented via media outlets to persuade people that fictional ideas are factual. Exposure to fake news can create memory errors that serve as the basis for inaccurate beliefs. Reported associations between false beliefs and related behaviors suggest that such beliefs could have serious consequences. For example, beliefs based on memory for inaccurate claims about the COVID-19 virus are associated with reduced compliance with health guidelines and reluctance to vaccination  has been shown in paradigms wherein participants read a fictitious news report that contained a critical piece of information that was subsequently corrected or not corrected. The CIE is observed when corrections reduce, but do not eliminate misinformation reliance. One suggestion for why corrections are not completely effective is that they sometimes repeat misinformation, thereby increasing the potential for familiarity to exert an unwanted influence. However, the view that familiarity will backfire has recently been contested because the available literature shows that such effects are rare and depend on experimental design and stimulus characteristics .The stimuli, data, and analysis scripts are available on the Open Science Framework (OSF) website: M = 19.23, SD = 1.32). Our stopping rule was to test as many participants as possible in approximately one semester given lab resources with the final sample being a multiple of the three experimental formats. We were primarily interested in how proactive effects of memory differed based on whether fake new details that were recalled during corrections were subsequently recalled in Phase 3. However, we also conducted a sensitivity analysis after collecting the data to estimate the statistical power to detect the smallest effect size (odds ratio [OR]) of interest in Experiment 1. This corresponded to the difference in real news recall in the Control and Correction headline conditions (OR = 1.36). The analysis indicated that Experiment 1 had 71% power to detect an odds ratio of 1.36. A complete description of this analysis and its results can be found in Supplementary Information , ages 18–25 (The experiment used a 3 (Headline Type: Repetition vs. Control vs. Correction) within-participants design.We selected from various internet sources 45 headline pairs that each included one fake news headline and its real news correction. The headlines were taken from sources such as news center websites , well-known people , government statistics websites, and folk myth websites. The fake news headlines were factual errors, and the real news corrections included factual details that contradicted the errors. All fake news headlines were originally portrayed as being true. Each headline pair corresponded to a unique topic. For example, the topic on the US president who took the most vacation days included the fake news headline, “President Obama took fewer vacation days than any other recent president,” and the real news headline, “President Clinton took fewer vacation days than any other recent president.” When preparing the experiment, we were able to find 41 topics related to US events described in various news sources. To increase the number of topics, we also included 4 urban myths . To foreshadow, when preparing Experiment 2, we found enough US events to replace the urban myths used here. Both experiments showed the same patterns related to the main hypotheses of this study, thus mitigating concerns about headline-specific effects.For counterbalancing, we randomly divided the 45 headline pairs into three groups of 15. We rotated groups through Headline Type conditions, which created three experimental formats. Groups appeared equally often in each condition across participants. Note that Phase 1 included a mixture of fake and real news headlines depending on the condition, whereas Phase 2 included all real news headlines. Phase 1 included 30 headlines . Phase 2 included 45 headlines . We included control headlines as a contrast condition against which to assess proactive effects of fake news exposure on memory for real news headlines.President Obama” or the real news detail “President Clinton.”Phase 3 included 45 questions corresponding to Phase 2 headlines that could be answered with either fake or real news details. For example, the question “Which recent president took the fewest vacation days?” could be completed with the fake news detail “Participants completed the experiment individually in a quiet room with an experimenter present. Stimuli were presented electronically using E-Prime 2.0 software  to 6 (True). The second prompt asked participants to rate their familiarity with the headline on a scale from 1 (Unfamiliar) to 6 (Familiar). Participants were encouraged to use the full range of the scale and made their ratings by pressing a number on the keyboard. The prompt color changed from white to yellow after participants entered their ratings. Each slide was followed by a blank screen for 250 ms. The purpose of including these judgments was to obtain baseline assessments of beliefs and familiarity for all items and to give participants a task that would keep them actively engaged.M = 1587 ms, 95% CI ) than Correction , t(2066) = 2.91, p = .01, but not Repetition  headlines, t(2066) = 2.03, p = .11. Durations for Correction and Repetition headlines were not significantly different, t(2066) = 0.88, p = .65.Prior to Phase 2, participants were told that some of the headlines they studied in Phase 1 were fake news and that corrections to those headlines would appear in Phase 2. They were also told that Phase 2 would include repetitions of real news headlines from Phase 1 and real news headlines that only appeared in Phase 2. Participants were told to study the headlines carefully for a later test and to indicate which headlines were corrections of Phase 1 fake news. During Phase 2, headlines appeared individually in a random order for 8000 ms each. Next, a prompt appeared below each headline asking participants to indicate if the headline corrected a headline that they studied in Phase 1. Participants made self-paced judgments, responding either Yes (1) or No (2) by pressing a number key. When they responded “Yes,” another prompt appeared above a text box wherein they typed only the fake news details from Phase 1 that were corrected in Phase 2 and pressed “Enter” to advance to the next headline. When they indicated “No,” the program advanced to the next headline. Note that the headlines stayed on the screen while participants made their responses. A mixed-effect model with Participants and Items as random intercept effects indicated that the estimated marginal mean of yes/no durations was significantly lower for Control  or No (2) by pressing a number key. When they indicated “Yes,” a prompt asked participants to type in the fake news details from Phase 1 that were corrected in Phase 2. When they indicated “No,” the program advanced to the next item. Participants were encouraged to respond accurately to all items and were allowed to pass when they could not think of a response.lme4 package  and response bias (c). We calculated these parameter estimates by computing hit and false alarm rates for each participant and using the dprime function from the psycho package . The significance level was α = .05.All models included Headline Type as a fixed effect as well as Participants and Items as random intercept effects to increase power  and fake news , t(1238) = 1.04, p = .30. In contrast, the model for belief ratings indicated that participants believed real news , t(1241) = 3.74, p < .001. Note that the familiarity ratings were modest relative to the maximum possible rating, indicating that participants knew little about many of the headlines before entering the experiment.Participants’ familiarity with and beliefs in Phase 1 headlines were compared for real and fake news headlines by fitting separate models to each measure with Headline Type as a factor. The model for familiarity ratings indicated that participants were comparably familiar with real news  = 159.32, p < .001. Correct recall was significantly greater for Repetition than Control and Correction headlines, smallest z ratio = 10.09, p < .001, and for Correction than Control headlines, z ratio = 2.48, p = .04, indicating overall proactive facilitation in memory for real news that corrected fake news, and thus, no overall familiarity backfire effect.Figure χ2(2) = 96.54, p < .001. Intrusions were significantly higher for Correction headlines than Repetition and Control headlines, smallest z ratio = 7.50, p < .001, and not significantly different for Repetition and Control headlines, z ratio = 1.24, p = .43. These findings show that fake news exposure in Phase 1 led to proactive interference when participants attempted to recall of real news details from Phase 2.Figure d’) and bias (c). In the signal detection analyses, the items in the Correction condition were treated as “old,” and the items in the Control condition were treated as “new.” We included Control headlines and excluded Repetition headlines, because Control headlines did not include features from Phase 1 and were thus more novel.To identify the role of detection of fake news corrections and recollecting that fake news had been corrected in cued recall accuracy in Phase 3, we first assessed participants’ awareness of and memory for such corrections in Phase 2 and Phase 3, respectively. We first computed the separate probabilities of participants indicating during Phase 2 that “yes” a headline was a correction and that during Phase 3 that “yes” the topic had earlier included a correction for all headline types. This provided a general indication of how well participants could discriminate Correction headlines from other headline types. Note that the extent to which participants incorrectly indicated that topics from the Repetition and Control conditions were associated with fake news corrections can be considered one index of response bias. Table χ2(2) = 1236.54, p < .001, showing that probabilities were significantly higher for Correction than Repetition and Control headlines, smallest z ratio = 29.80, p < .001, and for Repetition than Control headlines, z ratio = 8.59, p < .001. There was also a significant effect of Phase, χ2(1) = 8.71, p < .01, that was qualified by a significant interaction, χ2(2) = 37.93, p < .001. The response probabilities for Correction headlines were significantly higher in Phase 2 than in Phase 3, z ratio = 5.71, p < .001, for Control headlines were significantly higher in Phase 3 than in Phase 2, z ratio = 3.70, p < .001, and for Repetition headlines were not significantly different between phases, z ratio = 0.62, p = .54. Together, these results show that participants’ ability to discriminate headlines that were corrected from the other headline types diminished from Phase 2 to Phase 3. This discrimination difference was verified by fitting separate models to estimates for each signal detection parameter  = 4.00, p < .001, whereas the model for c indicated no significant difference in response bias between phases, t(94) = 0.78, p = .44.Awareness of and memory for corrections was first assessed by comparing the extent to which participants indicated that “yes” a headline topic had been Phase 2) or was (Phase 3) a correction  and for which fake news from Phase 1 was correctly recalled after the classification judgment. Classified as Correction + Fake News Not Recalled refers to Correction headlines that were classified as such and for which fake news from Phase 1 was not subsequently recalled. Not Classified as Correction refers to Correction headlines that were not classified as such . Since the observations across these three cells were not independent, we only statistically compared the probabilities of fake news recall between the two phases (Table χ2(1) = 38.66, p < .001. We did not compare fake news recall in the Repetition and Control headline conditions because participants rarely reported headline details that had not appeared earlier in the experiment (M ≤ .03).Next, we partitioned Correction headlines into three groups based on how participants classified them in Phase 2 and Phase 3  fitted to real news correct recall and fake news intrusions. We then examined how three levels of Classifications in Phase 3 for only when fake news was recalled in Phase 2  were associated with correct recall and intrusions using separate models for each. We conditioned the analyses on successful fake news recall during Phase 2 because those instances are of primary theoretical interest regarding whether retrieving earlier-studied headlines, would be associated with improved or impaired memory updating for real news headlines. Based on results from similar paradigms  = 28.82, p < .001, showing that correct recall was significantly higher for instances where corrections were classified as such and fake news was recalled (left green point) than when it was not recalled (left blue and red point), smallest z ratio = 3.59, p < .01. There was no significant difference in correct recall in the two cells for which fake news was not recalled in Phase 2, z ratio = 0.96, p = .60. Further, a model including the three types of Classifications in Phase 3 for when fake news was recalled in Phase 2 indicated a significant effect, χ2(2) = 90.77, p < .001, showing that recalling fake news was associated with enhanced recall of real news (right green point) compared to when fake news was not recalled in the two other cells (right blue point and red point), smallest z ratio = 7.56, p < .001. There was no significant difference in real news recall in the two cells for which fake news was not recalled in Phase 3, z ratio = 1.92, p = .13.Figure z ratio = 9.21, p < .001. Collapsing across the other conditional cells for which fake news was not recalled, failing to recall fake news was associated with proactive interference, as correct recall was significantly lower than for Control headlines, z ratio = 5.52, p < .001. These results showed that overall recall of corrections reflected a mixture of facilitation and interference for which the balance depended on the extent to which participants engaged recollection-based retrieval.To assess proactive effects of memory, correct recall was compared between these Phase 3 conditional cells and Control headlines. Recalling fake news was associated with proactive facilitation, as correct recall was significantly higher than for Control headlines, χ2(2) = 9.03, p = .01, showing that intrusions when fake news was recalled (left green point) than when corrections were classified as such but fake news was not recalled (left blue point), z ratio = 3.00, p < .01. There was no significant difference in intrusions for when corrections were not classified as such in Phase 2 (left red point) compared to the other classifications (left green and blue points), largest z ratio = 2.25, p = .06.Figure 2(1) = 0.62, p = .43. These instances were then collapsed and compared with Control headlines to examine proactive effects of memory when fake news was not recalled. The model indicated that average intrusion rate of these cells was significantly higher than the baseline rate for Control headlines, z ratio = 10.79, p < .001. This is consistent with the results from correct recall showing that failing to recall fake news was associated proactive interference in memory for real news headlines, which is a form of familiarity backfire.The proportions of intrusions for only fake news recall in Phase 2 were then examined based on whether fake news was not recalled in Phase 3 (right blue point and red point). Intrusions for which fake news was recalled were not analyzed because those errors occurred when participants reported the fake news twice in succession, once as being the correct headline detail and once as being the fake news headline detail. These instances of guessing were rare (1%), but the cell appearing in Fig. The results from Experiment 1 showed that correcting fake news increased correct recall for real news relative to when real news headlines appeared once in Phase 2 . These results are inconsistent with familiarity backfire accounts and are generally consistent with the integrative encoding account. However, conditional analyses revealed evidence for both familiarity backfire and enhanced updating that depended on whether participants could recall fake news and that it was corrected. These results extend those of Wahlheim et al.  and are Experiment 1 characterized proactive effects of memory when fake news was retrieved during corrections and subsequently recollected. The patterns showing that successful fake news retrieval during corrections was associated with proactive facilitation when fake news was recollected and proactive interference when fake news was not recollected generalizes earlier findings . As in Experiment 1, our stopping rule was to test as many participants as possible during one semester with the final sample size being a multiple of the total number of experimental formats, which was four (see below). A sensitivity analysis was conducted to estimate the observed power for detecting the smallest effect of interest, which was the difference in real news recall between the Control and Correction [Peers-Believe] condition. The present experiment had 99.50% power to detect an OR = 1.55  ages 18–27 (The experiment used a 4 (Headline Type: Repetition vs. Control vs. Correction [Peers-Believe] vs. Correction [Peers-Disbelieve]) within-participants design.https://osf.io/xnvrj/). For counterbalancing, headlines were rotated through conditions and appeared equally often in each condition across participants, as in Experiment 1. Phase 1 included 45 headlines . Phase 2 included 60 headlines ); 15 real news headlines that corrected Phase 1 fake news appearing earlier with an indicator showing that most peers disbelieved the headline (Correction [Peers-Disbelieve]); and 15 real news headlines that only appeared in Phase 2 (Control)]. Phase 3 included 60 questions that queried the critical details of Phase 2 headlines.The material set included 60 headline pairs comprising one fake news headline and its correction. This set included the original 41 items corresponding to current events from Experiment 1 and 19 new items about other current events. Based on an item analysis of the Experiment 1 data, we also clarified the wording of three items from the Experiment 1 set. The materials are available on the OSF , while the “do not believe” counter displayed an integer representing the remaining percentage . In the Peers-Disbelieve condition, this proportion was reversed between counters.M = .87) to provide a more even distribution of items among cells for the recall analyses conditioned on Phase 2 correction detection responses. This modification was especially useful for examining differences in fake news intrusion errors (see below).We selected moderate values instead of more extreme values because more extreme values seemed less plausible from a participant’s perspective. However, we acknowledge that social influence can occur when extreme values are used Kim, . For RepM = 1477 ms, 95% CI ) than Correction [Peers-Disbelieve]  headlines, z ratio = 2.79, p = .03. The estimate for Control headlines  was not significantly different than the estimates for Repetition, Correction [Peers-Believe] , Correction [Peers-Disbelieve] headlines, largest z ratio = 2.52, p = .06. Finally, there was no significant duration difference between the Correction [Peers-Believe] and Correction [Peers-Disbelieve] headlines, z ratio = 1.42, p = .49.Phase 2 was the same as in Experiment 1. A mixed-effect model with Participants and Items as random intercept effects indicated that the estimated marginal mean of yes/no durations was significantly lower for Repetition (χ2(3) = 364.47, p < .001. Correct recall was significantly greater for Repetition than the other three headline types, smallest z ratio = 13.78, p < .001. Correct recall was also significantly higher for Correction [Peers-Believe] and Correction [Peers-Disbelieve] headlines than Control headlines, smallest z ratio = 4.35, p < .001. There was no significant difference between the Correction [Peers-Believe] and Correction [Peers-Disbelieve] headlines, z ratio = 1.25, p = .59. Replicating Experiment 1, these results indicate overall proactive facilitation in memory for real news that corrected fake news, and thus, there is no overall familiarity backfire effect.Figure χ2(3) = 204.22, p < .001. Intrusions were significantly higher for both types of Correction headlines than the other headline types, smallest z ratio = 8.21, p < .001. There were no significant differences between Correction [Peers-Believe] and Correction [Peers-Disbelieve] headlines, z ratio = 0.45, p = .97, but intrusions were significantly higher for Control headlines than Repetition headlines, z ratio = 5.40, p < .001. Replicating Experiment 1, these findings show that fake news exposure in Phase 1 led to proactive interference when participants attempted to recall of real news details from Phase 2.Figure χ2(3) = 2255.02, p < .001, showing that probabilities were significantly higher for Correction [Peers-Believe] and Correction [Peers-Disbelieve] than Repetition and Control headlines, smallest z ratio = 34.96, p < .001. There was no significant difference between the Correction [Peers-Believe] and Correction [Peers-Disbelieve] headlines, z ratio = 0.71, p = .89, or Repetition and Control headlines, z ratio = 0.83, p = .84. There was also a significant effect of Phase, χ2(1) = 99.55, p < .001, showing that probabilities were significantly higher in Phase 2 than in Phase 3. There was a significant interaction, χ2(3) = 48.80, p < .001, showing that the probabilities for Correction [Peers-Believe], Correction [Peers-Disbelieve], and Control headlines were significantly higher in Phase 2 than in Phase 3, smallest z ratio = 3.47, p < .001, and the probabilities for Repetition headlines were not significantly different between phases, z ratio = 1.87, p = .06. Collectively, these findings show that participants were less able to discriminate Correction headlines from other headline types in Phase 3 than in Phase 2.Following our approach in Experiment 1, we identified the role of detection of fake news corrections and recollecting that fake news had been corrected in cued recall accuracy by first assessing awareness of and memory for corrections in Phase 2 and in Phase 3  = 5.74, p = .02. In addition, the model for c indicated that estimates were significantly higher in Phase 2 than in Phase 3, F = 33.36, p < .001, showing that participants became more conservative across phases. The less conservative responding in Phase 2, which did not replicate results from the first experiment, may have occurred because corrections comprised half of the items here and only a third of the items in the first experiment. No other effects were significant, largest F = 0.02, p = .88.To verify this claim, separate models were fitted to the estimates from each signal detection parameter (right panel). The model for z ratio = 0.44, p = .97, and that fake news recall for those conditions was significantly higher in Phase 2 than Phase 3, smallest z ratio = 5.59, p < .001. Participants rarely reported details from the fake news headlines that had not appeared in the experiment for Repetition and Control headlines (M ≤ .02).Also following our approach in Experiment 1, we next partitioned the classifications for Correction headlines in each phase into three types Table  to assesThe role of detecting corrections and recalling fake news on the proactive effects of memory was examined using the same approach as in Experiment 1.χ2(2) = 41.55, p < .001, no significant effect of Headline Type, χ2(1) = 0.25, p = .62, and no significant interaction, χ2(2) = 1.81, p = .40. Pairwise comparisons showed that correct recall was significantly higher when corrections were classified as such and fake news was recalled (left green point) than the other two classifications (left blue and red point), smallest z ratio = 2.52, p = .03. Correct recall was also significantly higher when corrections were classified as such (left blue point) than when they were not (left red point), z ratio = 3.59, p = .001.Figure χ2(2) = 80.45, p < .001, no significant effect of Headline Type, χ2(1) = 1.24, p = .27, and no significant interaction, χ2(2) = 0.48, p = .79. Pairwise comparisons showed that correct recall of real news was significantly higher when fake news was recalled (right green points) than when corrections were classified as such and fake news was not recalled (right blue points) and when corrections were not classified as such (right red points), smallest z ratio = 6.03, p < .001. There was no significant difference in correct recall in the two cells for which fake news was not recalled (right blue and red points) z ratio = 1.10, p = .52.Further, a model including the three types of Classifications in Phase 3 when fake news was recalled in Phase 2 and Headline Type as factors indicated a significant effect of Classification, t ratio = 12.04, p < .001. Collapsing across the other conditional cells for which fake news was not recalled, failing to recall fake news was associated with proactive interference, as correct recall was significantly lower than for those Correction than Control headlines, t ratio = 3.88, p < .001. Replicating Experiment 1, these results show that overall recall of corrections reflected a mixture of facilitation and interference for which the balance depended on the extent to which participants engaged recollection-based retrieval. However, a novel finding here was that higher memory accuracy for real news when corrections were classified as such but no fake news details were retrieved relative to not classifying a correction.To examine the proactive effects of fake news exposure on subsequent memory for corrections, we first collapsed the Correction [Peers-Believe] and Correction [Peers-Disbelieve] conditions into a single Correction condition because there were no differences in conditional recall between these two conditions. We then compared conditional correct recall in the Correction headline conditions with the Control condition. A model including Headline Type as a factor with the three classification types for corrections in Phase 3 when fake news was recalled in Phase 2 and Control headlines as the levels indicated that recalling fake news was associated with proactive facilitation, as correct recall was significantly higher for those Correction than Control headlines, χ2(2) = 11.65, p < .01, showing that there were fewer intrusion errors for when corrections were classified and fake news was not recalled (left blue points) than when corrections were not classified as such (left red points), z ratio = 3.43, p < .01. There was no significant difference in intrusions for when corrections were classified as such and fake news was recalled (left green points) compared to the other classifications (left blue and red points), largest z ratio = 1.93, p = .13. There was no significant effect of Headline Type, χ2(1) < 0.01, p = .97, and no significant interaction, χ2(2) = .92, p = .63.Figure χ2(2) = 2.20, p = .14, no significant effect of Headline Type, χ2(1) = 0.13, p = .72, and no significant interaction, χ2(1) = 0.20, p = .66. These instances were then collapsed and compared with Control headlines to examine proactive effects of memory when fake news was not recalled. The model indicated that average intrusion rate of these cells was significantly higher than the baseline intrusion rate for Control headlines, t ratio = 18.14, p < .001. As in Experiment 1, the overall patterns here are consistent with the results from correct recall showing that failing to subsequently recall fake news was associated proactive interference in memory for real news headlines, which is a form of familiarity backfire. However, the fewer intrusions associated with classifying a correction but not recalling fake news are a novel finding in this study.The model for intrusions for when fake news details were recalled in Phase 2 conditioned on whether they were not subsequently recalled (right red and blue points) indicated no significant effect of Classification, We also explored if the congruence between peer and participant beliefs influenced the encoding of fake news headlines in Phase 1 and its consequences for performance on memory measures in subsequent phases . First, the self-paced nature of the study phase could have contributed to memory differences for fake news corrections because participants spent slightly more time making judgments for those items in Experiment 1, and the headline remained on the screen until participants typed their recall response. This additional time could have been partly responsible for the proactive facilitation observed in overall real news recall. However, there was likely still a prominent role of integrative encoding in this recall advantage as an interference-based account would assume that more time spent encoding alternative associations when fake news was recalled would lead to subsequent proactive interference in overall recall. Future studies could control for encoding time differences by holding the duration constant across headline types and omitting Phase 1 recall following correction detection judgments. However, the average duration of correction detections was comparable between control and correction headlines in Experiment 2, providing evidence against a differential encoding account of proactive facilitation in overall recall.Second, the stimulus materials and presentation format of corrections provided ecologically valid information content and a way to assess the downstream consequences of recalling fake news during corrections. Although consumers may read news headlines as short declarative sentences, as in the present study, headlines sometimes appear with images, and news is sometimes delivered in modalities with unfolding temporal structures, such as videos or podcasts. Additionally, it is not often that an outside source tells someone to read headlines and compare the content to existing memories for earlier-read headlines.To conclude, this study extends the line of research highlighting the importance of recollection-based retrieval following retrieval of fake news details during corrections for subsequent updating memory for real news details. It allowed us to test competing predictions from leading theoretical perspectives on memory and belief updating, namely the familiarity backfire and integrative encoding accounts. The present findings implicated roles for mechanisms from both accounts along with a critical moderating role for recollection-based retrievals. These findings suggest that successfully retrieving fake news details when reading real news headlines can promote the comparisons necessary to encode associations and support later recollection of news headlines and their veracity. Future interventions may be improved by considering how interactions between retrieval strategies engaged during encoding and the inclusion of overlapping features between real and fake news headlines can promote enduring memory representations upon which beliefs can be based.Additional file 1: Supplementary Information.


Serratia (now part of the Yersiniaceae family) where data are limited to a few genomes of the species marcescens. This study describes the detection, in silico, of CRISPR loci in 146 Serratia complete genomes and 336 high‐quality assemblies available for the species ficaria, fonticola, grimesii, inhibens, liquefaciens, marcescens, nematodiphila, odorifera, oryzae, plymuthica, proteomaculans, quinivorans, rubidaea, symbiotica, and ureilytica. Apart from subtypes I‐E and I‐F1 which had previously been identified in marcescens, we report that of I‐C and the I‐E unique locus 1, I‐E*, and I‐F1 unique locus 1. Analysis of the genomic contexts for CRISPR loci revealed mdtN‐phnP as the region mostly shared . Three new contexts detected in genomes of rubidaea and fonticola (puu genes‐mnmA) and rubidaea (osmE‐soxG and ampC‐yebZ) were also found. The plasmid and/or phage origin of spacers was also established.The Clustered Regularly Interspaced Short Palindromic Repeats and CRISPR‐associated proteins (CRISPR–Cas) system of prokaryotes is an adaptative immune defense mechanism to protect themselves from invading genetic elements . Studies that describe the genetic organization of these prokaryotic systems have mainly reported on the Enterobacteriaceae family . For some genera, data on CRISPR–Cas systems remain poor, as in the case of  This study has contributed to extending the general knowledge of the CRISPR–Cas systems, particularly in Serratia, a ubiquitous genus detected in different environments. Despite this reclassification, data on CRISPR–Cas systems remain mainly limited to genera of the Enterobacteriaceae family  is a defense mechanism for bacteria and archaea against the invasion of bacteriophages and selfish genetic elements such as plasmids. Since their discovery around 15 years ago  and hosts  where they may play different roles ranging from opportunistic pathogens to symbionts  for human bacteremia are liquefaciens and odorifera  of subtype I‐C, the presence of unique loci, and detailed genomic contexts of CRISPR loci. The plasmid and/or phage origin of spacers was also assessed.In this study, 146 Serratia, has undoubtedly contributed to an expanded knowledge of these systems.The discovery of CRISPR–Cas systems has allowed the development of new technology tools in the bioengineering field  up to December 12, 2020  up to December 12, 2020, we selected the high‐quality assemblies  that have been included in the study.One hundred and forty‐six Serratia strains AS12 (NC_015566.1), FGI94 (NC_020064), FS14 (NZ_CP005927), SCBI (NZ_CP003424), YD25 (NZ_CP016948), and DSM21420 (GCA_000738675) were reclassified as reported by Sandner‐Miranda et al.  were recovered (when available) from GenBank or related articles. a et al. , Sandnera et al. . In the S. marcescens strains reported as secondary symbionts in the Red Palm Weevil (RPW) Rhynchophorus ferrugineus  (Coleoptera: Curculionidae)  harboring CRISPR loci in genome assemblies (unpublished) of four 2.2++ database. CRISPR arrays recorded by CRISPR–Cas++ were assigned to Levels 1–4 based on the criteria required to select the minimal structure of putative CRISPR as reported by Pourcel et al. (http://omics.informatics.indiana.edu/CRISPRone/)  and CRISPR arrays only for complete genomes were retrieved from the CRISPR–Casl et al. . Level 12.3http://crispr.otago.ac.nz/CRISPRBank/index.html) and CRISPR–Cas++ to assign, based on identity with known consensus DRs  was performed through the CRISPRCasdb spacer database at the CRISPRCas++ site (https://crisprcas.i2bc.paris-saclay.fr/MainDbQry/Index). Phagic and/or plasmidic origin of matching protospacers were searched at the CRISPRTarget site       Edgar, , 2004b. 33.1Serratia complete genomes was explored for the presence of cas gene clusters and/or CRISPR arrays. Most of the genomes (134) were reported as known species: ficaria (1), fonticola (7), grimesii (1), inhibens (1), liquefaciens (7), marcescens (87), nematodiphila (1), plymuthica (11), proteomaculans (2), quinivorans (2), rubidaea (8), symbiotica (4), ureilytica (2). The remaining 12 genomes were of unidentified species and, from here on, they will be referred to as Serratia sp. (Supporting Information: Table The CRISPR–Cas systems or only CRISPR arrays (orphan array) were detected in 35 complete genomes (24%) of which 17 harbored a CRISPR–Cas system, while 18 harbored orphan arrays. Some complete genomes characterized by the same cas gene set subtype and identical numbers of both CRISPRs and spacers were assumed as multiple records of the same genome (Table rubidaea)  and six I‐F1   and I‐F1 unique locus 1 .A collection of 146 me Table  and dist) Figure , one I‐E) Figure . The remcas3‐cas8e genes spaced by ~600 nt while the I‐F1 unique locus 1 had the cas3‐cas8f1 genes separated from each other by ~400 nt , fonticola (6), grimesii (2), liquefaciens (3), marcescens (295), nematodiphila (2), odorifera (2), oryzae (1), plymuthica (4), proteomaculas (1), rubidaea (2), symbiotica (1), ureilytica (1), and Serratia sp. (15)     an additional unique locus of the subtype I‐E, identical to I‐E* previously reported by Shen et al.  of a gene harboring the WYL domain which encodes for a potential functional partner of the CARF (CRISPR–Cas Associated Rossmann Fold) superfamily proteins   Figure , two sub) Figure , and thr) Figure . The I‐En et al. , was detn et al. . The loc2) Table .3.2fonticola (4), inhibens (1), marcescens (19), plymuthica (5), rubidaea (15), and Serratia sp. (4). Twenty‐three arrays were orphans and detected in genomes of marcescens (8), plymuthica (4), symbiotica (1), nematodiphila (1), rubidaea (5), and Serratia sp. (4)   has previously been described in S. marcescens strains isolated as a secondary symbiont of RPW and in other marcescens complete genomes available in the NCBI database , C (osmE‐soxG), and D (ampC‐yebZ) were shared by 11, four, and six genomes, respectively; context B by genomes of species fonticola (2), rubidaea (7), and Serratia sp. (2); C and D only by rubidaea genomes. For context D, assignment to rubidaea was assumed for the strain JUb9 (see above). The contexts E (nrdG‐bglH) and F (sucD‐vasK) were both identified in the single genome of S. oryzae strain J11‐6; while G (gntR‐cda) and H (gutQ‐queA) in genomes of the Serratia sp. Ag1 and S. symbiotica CWBI‐2.3, respectively  with the context C associated with the cas gene set of subtype I‐C.The 35 CRISPR‐positive complete genomes and 28 of the 46 CRISPR‐positive high‐quality assemblies were analyzed to identify possible shared genomic contexts. Eight different genomic contexts, named from A to H, were identified. Contexts A to D  or a metabolic burden  to which CRISPR–Cas systems represent a unique adaptative immunity defense mechanism. Studying the presence/absence of CRISPR–Cas systems and their features in different genera of families is a relatively new scientific approach to investigation to gain data on the evolution of these systems and their role played during the bacterial lifetime   or a misrepresentative distribution of the available genomes in databases, remains to be established.Data on CRISPR loci in mdtN‐phnP previously reported in the species marcescens and now further extended to those of grimesii, inhibens, nematodiphila, plymuthica, and rubidaea. Three new possible contexts were also identified: one (puu genes‐mnmA) shared by genomes of rubidaea and fonticola; and two (osmE‐soxG and ampC‐yebZ) detected in those of rubidaea. The context osmE–soxG might be closely linked to the cas gene set of subtype I‐C  showed the predominant presence of subtype I‐E and the rare coexistence of subtypes I‐E and I‐F1 in the same genome  in only three Serratia genomes, confirms that the coexistence of these subtypes is not frequent. It is also important to note that the only Serratia strain harboring a type III system reported by Medina‐Aparicio et al. , I‐E* (characterized by the cas6e translocation between cas7 and cas11) and I‐F1 unique locus 1 (characterized by 0.4 kb cas3/cas8f1 intergenic sequence) are, to the best of our knowledge, the first ever detected in Serratia. Similar or identical architectures of I‐E unique locus 1, I‐E*, and I‐F1 unique locus 1 have been reported for other bacteria genera: a similar architecture to I‐E unique locus 1 has been described in Escherichia coli (IGLB fragment) where the cas3/cas8e intergenic sequence was ~0.4 kb  strains , where the cas3/cas8f1 intergenic sequence was ~0.1 kb , orphans (29.4%) and aliens (10.2%) arrays were also detected  within the I‐E branch and thus might represent and/or anticipate a distinct clonal expansion of an I‐E sub‐population Figure .Cas systems is constantly expanding due to studies on newly available genomic sequences or genomic sequences not yet explored. The CRISPR–Cas systems classification is thus continuously updating also in light of their possible applications. Indeed, the CRISPR–Cas technology has undoubtedly revolutionized systems of genome editing with a wide range of potential industrial and biomedical applications. Other, more recent genome‐editing tools are based on methods that make use of the Cas9 protein  relevant in environmental and clinical fields. Moreover, the detection of the same subtype of cas‐gene sets in different Serratia species and other genera highlights the open question of the molecular mechanisms yet to be identified that have allowed intra‐ and inter‐species spread.Our study has contributed to expanding knowledge of the variability and distribution of CRISPR systems in the Maria Scrascia: Conceptualization ; investigation ; methodology ; writing – original draft ; writing – review and editing . Roberta Roberto: Formal analysis ; investigation . Pietro Daddabbo: Formal analysis . Yosra Ahmed: Data curation . Francesco Porcelli: Conceptualization . Marta Oliva: Investigation . Carla Calia: Investigation . Angelo Marzella: Investigation . Carlo Pazzani: Methodology ; supervision ; writing – original draft ; writing – review and editing .None declared.None required.Serratia genome assemblies.Table S1: List of Click here for additional data file.Table S2: Spacer analyses.Click here for additional data file.Figure S1: Phylogenetic tree of 16S rRNA gene.Click here for additional data file.


We aimed to test whether the prediction of presurgical metabolic syndrome for postsurgical survival outcomes of gastric cancer hinges upon cigarette smoking status.This study is a part of the ongoing Fujian prospective investigation of cancer (FIESTA) study. Patients with gastric cancer received radical resection of primary gastric cancer between January 2000 and December 2010, with the latest follow‐up ended in December 2015. The 1:1 propensity score matching analysis was adopted to balance confounders between smokers and never‐smokers. Effect‐size estimates are expressed as hazard ratio (HR) with 95% confidence interval (CI). Model performance was evaluated using the Hosmer and Lemeshow test and 10‐fold cross‐validated area under the receiver operating characteristic curve (AUROC). Statistical analyses were completed with SAS software (v9.4).p < 0.001), but not in never‐smokers. Relative excess risk due to interaction was estimated to be 2.43 (95% CI: 0.40–4.45). After constructing a risk assessment score, one unit increment was associated with 10% reduced risk of gastric cancer‐specific mortality , with 10‐fold cross‐validated AUROC being 0.82 (95% CI: 0.74–0.92).Total 2779 patients with gastric cancer were analyzed, including 2223 smokers and 556 never‐smokers. Median follow‐up time was 45.6 months. Cigarette smoking was not associated with postsurgical survival differences. Presurgical metabolic syndrome complication was significantly associated with increased gastric cancer‐specific mortality in smokers , and Kaplan‐Meier curves by score groups in patients overall (B) and by cigarette smoking status (panel C in smokers and panel D in never‐smokers). MST, median survival time. Group 1 refers to patients with a total score ranging from 50 to 69; group 2 refers to patients with a total score ranging from 70 to 82; group 3 refers to patients with a total score ranging from 83 to 92. Additionally, given the facts that the development and progression of gastric cancer involve multiple factors and contribution of individual factors is likely to be small, we attempted to construct a risk scoring system to further assess survival outcomes.22.1The present study used data from the ongoing FIESTA study. The FIESTA study was initiated by Fujian Provincial Cancer Hospital in January 2000, with the aims to optimize presurgical risk factors and predict cancer‐specific mortality of common digestive system malignancies, including esophageal cancer, gastric cancer, and colorectal cancer.2.2Total 3012 patients with gastric cancer who received radical resection of primary gastric cancer for the first time, underwent no presurgical and postsurgical chemotherapy/radiotherapy, and were consecutively recruited from January 2000 to December 2010, with the latest follow‐up ended on December 31, 2015, were eligible for inclusion, as previously described.2.3Primary cancerous and adjacent normal tissue samples were collected during radical resection of primary gastric cancer, and they were fixed in 10% neutral‐buffered formalin and paraffin‐embedded using standard procedures. Presurgical biopsy and postsurgical pathological examination were used to confirm clinical diagnosis of gastric cancer.2.4Data were collected prospectively. Follow‐up interval was 6 months to 12 months after radical resection of primary gastric cancer. Patients who failed to appear or missed appointments at scheduled time were contacted them by phone calls or postal mails. All patients were followed up from initial admission after the surgery since January 2000 to death or last follow‐up visit before December 31, 2015, whichever occurred first.2.52; (ii) hyperglycemia: fasting blood glucose ≥6.1 mmol/L or 2‐h plasma glucose ≥7.8 mmol/L or previously diagnosed diabetes; (iii) hypertension: systolic/diastolic blood pressure ≥ 140/90 mm Hg or antihypertensive therapy; (iv) dyslipidemia: triglycerides ≥1.7 mmol/L or high‐density lipoprotein cholesterol (HDLC) <0.9 mmol/L in males or <1.0 mmol/L in females.In this study, metabolic syndrome was defined according to the criteria set forth by the Chinese Diabetes Society in 2004.2.6All patients were invited to complete a self‐designed questionnaire to obtain demographic data at baseline, including age, gender, ABO blood type, smoking status, drinking status, and family cancer history. Recorded age referred to age at the time of initially receiving radical resection of primary gastric cancer. Smoking status was categorized as never, former, and current smoking, with the latter two categories combined as ever smoking. Similarly, drinking status was categorized as never, former, and current alcohol drinking, with the latter two categories combined as ever drinking. Family cancer history was recorded if one or more direct relatives are diagnosed with cancer (except non‐melanoma skin cancer) within three generations. Body height and weight were taken on patients in light clothing and without shoes, and they were used to calculate BMI. Blood pressure was measured in a seated position strictly according to the standard protocol recommended by the American Heart Association.Fasting (overnight at least 8 h) venous blood samples were collected at the morning of receiving radical resection of primary gastric cancer. Serum concentrations of glucose, plasma triglycerides, and total cholesterol, HDLC, and low‐density lipoprotein cholesterol (LDLC) were measured using standard procedures. An automated glucose oxidase method was used to assay fasting blood glucose.In addition, routine blood indexes were recorded, including neutrophil, lymphocyte, monocyte, eosinophil, basophil, white blood cell count, red blood cell count, hemoglobin, red cell distribution width, and platelet count by the SYSMEX XE‐2100 Automatic Blood Cell Analyzer . Interval from blood drawing to biochemical detection was less than 4 h according to standard procedures. Based on these indexes, neutrophil‐to‐lymphocyte ratio (NLR), platelet‐to‐lymphocyte ratio (PLR), lymphocyte‐to‐monocyte ratio (LMR), and monocyte‐to‐red blood cell count ratio (MRR) were derived accordingly.Clinicopathologic characteristics including TNM stage  as the dependent variable. Variables that were selected based on clinical judgment and statistical significance in univariate comparisons were incorporated into propensity models, including age, gender, drinking, BMI, and family cancer history. An 1:1 propensity score matching analysis was then conducted between the smoking and non‐smoking groups based on derived propensity score. Those with closest scores were selected as pairs. After propensity score matching, hazard ratio (HR) with 95% confidence interval (CI) for gastric cancer‐specific mortality was estimated using adjusted and unadjusted Cox proportional hazard models. Proportional hazards assumption was tested by Schoenfeld residuals.http://c2s2.yale.edu/software/stree/). The risk score for each prognostic factor was calculated as dividing 173‐month survival rate by 10, according to the method recommended by Sarah Douglas and colleagues.Variables that showed statistical significance in both smokers and never‐smokers were chosen to develop a risk scoring system for gastric cancer mortality prediction. The optimal cut‐offs for continuous variables were selected based on survival tree analysis using the STREE program  and then tested on 1/10 of the remaining data (testing set). Then, a 10‐fold cross‐validated prediction model was performed to estimate area under the receiver operating characteristic curve (AUROC).Statistical analyses were completed by the SAS software, version 9.4 (SAS Institute Inc.), unless otherwise indicated.33.1At the end of follow‐up, there were 1331 (44.19%) deaths from gastric cancer and 1681 survivors. Median follow‐up time of all patients was 44.1 months (range: 1.1 to 183.3 months). Only 2779 patients had complete data on cigarette smoking status, including 2223 smokers and 556 never‐smokers. The baseline characteristics of study participants are shown in Table In view of striking differences in some baseline characteristics, such as gender composition and alcohol consumption, an 1:1 propensity score matching analysis was conducted between smokers and never‐smokers to balance confounders, and the baseline characteristics of study participants after the propensity score matching analysis are shown in Table 3.2After the propensity score matching analysis, Kaplan–Meier curves and Log‐rank tests failed to reveal any significance in cumulative survival rates for gastric cancer between smokers and never‐smokers. Schoenfeld residuals did not show major departures from the proportional hazards assumption.pinteraction < 0.001). Smokers with hypertension had 1.65‐fold increased risk for gastric cancer‐specific mortality , while never‐smokers with hypertension had 1.63‐fold increased risk . Smokers with obesity , dyslipidemia  or metabolic syndrome  showed a higher risk relative to smokers without, and interaction with never‐smokers was not significant for these variables  , 1.07 (95% CI: 0.85–1.34), and 4.08 (2.57–6.49), respectively, compared with never‐smokers without metabolic syndrome (Table Relative excess risk due to interaction (RERI) was estimated to be 2.43 (95% CI: 0.40–4.45), indicating that there was 2.43 relatively excess risk due to additive interaction    in patients overall and by cigarette smoking status Table . Moreovel) Table .Then, total score was classified into three groups based on their distributions among all study patients . Kaplan–Meier curves of the three groups for cumulative survival against follow‐up time in all study patients, smokers, and non‐smokers are displayed in Figure Survival rates were 22.90%, 60.26%, and 90.77% for group 1, group 2, and group 3, respectively. Relative to group 1, patients in group 2  and group 3  separately had 67% and 94% reduced risk of gastric cancer‐specific mortality after adjusting for confounders , prognostic nutritional index (PNI), and controlling nutritional status (CONUT) score,More recently, we and other researchers have found that metabolic syndrome was associated with an increased risk of gastric cancer‐specific mortality.The key finding of this study is the additive interaction between cigarette smoking and metabolic syndrome when predicting postsurgical gastric cancer‐specific mortality. Cigarette smoking is a well‐known and established risk factor for gastric cancer,Helicobacter pylori, an established risk factor for gastric cancer. Additionally, after the propensity score matching analysis, only 397 smokers and 397 never‐smokers were assessable in this study, which led to loss of statistical power. Fourth, only patients who received radical resection of primary gastric cancer, but no presurgical and postsurgical chemotherapy/radiotherapy were enrolled, which might limit the extrapolation of our findings to general patient groups. Additionally, our findings were exclusively derived from a southern Chinese population, and additional replications in other ethnic or racial groups are critical. Fifth, more male patients than female patients were enrolled in this cohort, which might restrict extrapolation to the more general populations.Besides the clear strengths of this study, including the prospective design, large sample size, and long follow‐up, several limitations should be acknowledged. First, we recruited patients from a single center, which may limit the generalizability of our findings, although it facilitates consistency of evaluation and intervention. Second, we had data on cigarette smoking only, and other important smoking‐related indexes, such as pack‐years smoked and smoking cessation were not available. Additionally, information on perisurgical factors such as surgical procedure, lymphadenectomy and reconstruction, as well as subsequent management of metabolic syndrome after radical resection of primary gastric cancer was not available for us, which might introduce a prediction bias. Third, although we adopted the propensity score matching analysis to balance baseline confounders between smokers and never‐smokers, there are still other confounders without taking into consideration, such as 5Our findings indicate that the prediction of presurgical metabolic syndrome for gastric cancer‐specific mortality is more evident in smokers. Importantly, we have constructed a risk scoring system, which exhibited a strong discriminatory capability to predict postsurgical survival outcomes in patients with gastric cancer. Although this scoring system needs external validation, it has provided evidence base for future personalized prediction and helped risk‐stratify gastric cancer patients who might experience serious postsurgical consequences.DH, WN planned and designed the study, and directed its implementation; FP, DH, XZ drafted the protocol; DH, XZ obtained statutory and ethics approvals; DH, XZ contributed to data acquisition; XD, XZ, FP, WN conducted statistical analyses; XZ, XD, WN, DH, FP, XZ had access to all raw data; FP, DH, FM, WN did the data preparation and quality control; XZ, FM, WN wrote the manuscript. All authors read and approved the final manuscript prior to submission.There are no commercial interests or sources of financial or material support to declare.The authors declare that they have no competing interests, and all authors should confirm its accuracy.All participants provided their written informed consent, and this study was approved by the Ethics Committee of Fujian Provincial Cancer Hospital.Appendix S1Click here for additional data file.


Leishmania, Borrelia, and Anaplasma spp.; however, Leishmania PCR (splenic and bone marrow aspirates), and paired serological titers for Borrelia and Anaplasma were negative. Consequently, initial serological results were considered to be false positive because of paraproteinemia‐associated assay interference. Chemotherapy  was initiated, but the dog was euthanased 30 days later because of the development of pericardial effusion. This is a report of spurious serological (and other laboratory) results occurring secondary to monoclonal gammopathy in a dog.An 11 year old female‐neutered Labrador presented for facial swelling. Clinicopathological abnormalities included hyperglobulinemia, azotemia, hypercalcemia, nonregenerative anemia, thrombocytopenia, and spurious hypoglycemia. Normoglycemia was subsequently confirmed using a cage‐side analyzer . Serum and urine protein electrophoresis documented monoclonal (immunoglobulin M) gammopathy with Bence‐Jones proteinuria. Computed tomography imaging revealed a monostotic osteolytic bone‐lesion, and bone marrow cytology and histopathology documented plasmacytosis with multiple myeloma oncogene 1 / interferon regulatory factor 4 positivity, consistent with multiple myeloma. Infectious disease testing initially indicated seropositivity for  CTcomputed tomographyELISAenzyme‐linked immunosorbent assayHPFhigh‐powered fieldIFATimmunofluorescence antibody testIgMimmunoglobulin MIRF‐4interferon regulatory factor 4MMmultiple myelomaMUM‐1multiple myeloma oncogene 1PEGpolyethylene glycolPOCUSpoint of care ultrasoundRIreference intervalUPCurine protein : creatinine1Hyperglobulinemia and monoclonal gammopathy in dogs is most commonly seen secondary to either neoplasia  or infectious diseases (such as ehrlichiosis and leishmaniasis),Circulating paraproteins in human patients cause spurious laboratory results,2An 11 year old female neutered Labrador was presented for further investigation of a 48‐hour history of acute onset left‐sided facial swelling. Investigations performed by the referring veterinarian documented thrombocytopenia, anemia, hematuria, and pain when opening the mouth. The dog was prescribed 10 mg/kg paracetamol  twice daily and 0.1 mg/kg meloxicam  once daily before referral. The dog had no pertinent medical history, was up‐to‐date with vaccinations and anthelmintic prophylaxis, and had no history of travel outside the UK.On presentation, the dog was bright with no evidence of dehydration. Examination documented left‐sided facial swelling, trismus and pain when opening the jaw. An ulcerated mass‐lesion was present at the left oral commissure. Ocular examination documented multifocal retinal hemorrhages, and systolic blood pressure (via Doppler interrogation) was 155 mm Hg, consistent with prehypertension and low risk of target‐organ damage.9/L  with thrombocytopenia 78 × 109/L , and plasma protein above the reference range 110 g/L . Coagulation profile was within normal limits. Serum biochemistry  documented elevated urea 50 mg/dL , creatinine 1.8 mg/dL , and phosphate 5.7 mg/dL , which coupled with isosthenuric urine (USG 1.015) was consistent with renal azotemia. It also documented hyperglobulinemia 76 g/L  with low‐normal albumin 25 g/L , and hypercalcemia  13 mg/dL  with normal ionised calcium 5.2 mg/dL . The most clinically relevant biochemical abnormality was hypoglycemia below the limit of quantification <12.6 mg/dL , initially documented via hexokinase assay. However, normoglycemia with an average of 153 mg/dL (lowest reading 139 mg/dL) was subsequently documented using a cage‐side coulometric assay analyzer , suggesting initial analytical error. Electrolytes, hepatobiliary and pancreatic markers, C‐reactive protein, cholesterol, and triglyceride were within normal limits. Urinalysis collected via cystocentesis documented hematuria 150 cells per high‐powered field (hpf) , pyuria 8 cells/hpf , and proteinuria with a urine protein : creatinine (UPC) ratio of 3.52  but no bacteriuria, and negative culture. Serum and urine protein electrophoresis documented a monoclonal immunoglobulin M (IgM) gammopathy and Bence‐jones proteinuria Angiostrongylus vasorum, Ehrlichia canis/ewingii, and Dirofilaria immitis, but indicated positive antibody detection for Borrelia burgdorferi and Anaplasma phagocytophilum/platys. Evaluation of acute and convalescent serological titers to B burgdorferi and A phagocytophilum/platys were subsequently performed but both were negative; , suggesting a previous analytical error. Quantitative ELISA for detection of IgG against Leishmania species (IDEXX Laboratories Ltd.) was also positive with a titer of 53.5 enzyme units ; however, immunofluorescence antibody test (IFAT) serology was inconclusive and subsequent Leishmania PCR on splenic and bone marrow aspirates were negative.Infectious disease testing was performed using cage‐side enzyme‐linked immunosorbent assay (ELISA) kits . These were negative for Abdominal ultrasonography was performed under sedation. The hepatic parenchyma contained a small rounded hypoechoic nodule (0.89 cm) but was otherwise normal size and shape. There were multiple small rounded hypoechoic nodules throughout the splenic parenchyma, but no other abnormalities were detected on the remainder of the abdominal ultrasound examination. Ultrasound‐guided fine needle aspiration of the liver and spleen documented vacuolar hepatopathy, and high numbers of well‐differentiated plasma cells, respectively.A contrast‐enhanced computerized tomographic (CT) scan of the head, neck, and chest was performed under general anesthesia. This identified a small, ill‐defined, rounded, poorly contrast‐enhancing swelling on the internal aspect of the left cheek. There was no evidence of local invasion or regional lymphadenopathy, and the lesion did not extend to the ipsilateral tonsil. Thoracic CT identified a large well‐defined heart‐base mass , located ventral to the aortic arch and brachiocephalic trunk, within the pericardium Figure . AlthougEchocardiography documented a large, well‐circumscribed heart‐base mass of heterogeneous echogenicity, located between the aorta and the right atrium within the pericardial space, with no evidence of invasion of the surrounding structures. The heart was otherwise structurally and functionally normal.Bone marrow cytology and histopathology Figure  documentThe neutrophilic stomatitis was treated with 20 mg/kg potentiated amoxicillin  twice daily whilst awaiting test results, which resulted in resolution of the left‐sided facial swelling and associated discomfort. After obtaining a diagnosis of multiple myeloma, chemotherapy was initiated with 0.1 mg/kg melphalan  once daily and 0.5 mg/kg prednisolone  once daily, with 1 mg/kg alendronic acid  once daily because of concurrent hypercalcemia.There was clinical and biochemical improvement after initiation of treatment, with resolution of the facial swelling and reduction in serum globulin concentration to 57 g/L  after 10 days. The dog represented 4 weeks later with a 24‐hour history of progressive exercise intolerance, tachycardia, and pulse deficits. A thoracic Point‐Of‐Care Ultrasound (POCUS) was performed, identifying a large volume of anechoic pericardial effusion causing collapse of the right atrium and tamponade. The heart‐base mass was thought to be grossly unchanged from the previous thoracic ultrasound. Given the clinical deterioration, the owners opted not to pursue further investigation and elected for euthanasia. A post‐mortem examination was offered but declined.3The pitfalls of paraproteinemia‐induced spurious laboratory results are well known and recognized in human medicine, leading to clinically impossible results,Leishmania assay in the described case could be associated with the paraprotein cross‐linking the antigen with the conjugate antibody,Along with spurious chemistry results, false‐positive serological results occur in humans secondary to paraproteinemia, both because of hematological malignancies (such as multiple myeloma) and secondary to benign age‐related increases in immunoglobulin levels. This includes false‐positive Galactomann antigen tests for aspergillosis,A phagocytophilum/platys, B burgdorferi, and Leishmania spp. therefore prompted further evaluation using confirmatory tests. There is no gold standard diagnostic test for leishmaniasis.Leishmania spp.In this case, although there was no history of foreign travel, infectious disease was still initially considered a possible etiology, given that anaplasmosis,Although the SNAP 4dx plus test (IDEXX Laboratories Ltd) is highly sensitive, false‐positives can also occur, particularly in regions with low disease prevalence.Because of the known effect of paraproteinemia causing spurious laboratory results in human medicine, multiple protocols have been developed to try and address this and improve test accuracy. Some authors propose serial dilution of samples with saline until concordant results are achieved,An alternative method to avoid interference includes deproteination of the sample.A limitation in this case was the lack of cytological or histopathological evaluation of the heart‐base mass, and as such we cannot conclude the underlying etiology of this lesion and whether it was an incidental comorbidity in an older animal (such as chemodectoma), or a rare presentation of extramedullary plasmacytoma. Mediastinal plasmacytoma occurring in people with multiple myeloma is very rarely documented, with only 20 cases in the literature,The authors declare no conflict of interest.The authors declare no off‐label use of antimicrobials.Written owner consent was obtained at the time of animal admission.The authors declare human ethics approval was not needed for this study.


Currently, a new wave of coronavirus disease 2019 (COVID‐19) pandemic appears worldwide with the omicron variant dominating. Compared to the other variants of SARS‐CoV‐2, omicron is more transmissible and with a shorter incubation period than the other variants such as delta, albeit with a reduced impact on lower respiratory tract and rates of hospital admission.Acute and chronic infections are the common complications of diabetes. Although there is insufficient evidence to confirm that patients with diabetes are more susceptible to COVID‐19 than the general population, multiple studies have shown that the patients' underlying condition such as diabetes was associated with an increased risk of poorer clinical outcomes including mortality for these patients.Here, we propose this expert suggestion to provide necessary recommendation for the comprehensive multifactorial management of diabetes patients during the current COVID‐19 pandemic, especially under the current omicron wave.For patients with well‐controlled diabetes, COVID‐19 vaccines should be given in time. Moreover, after evaluation, booster vaccination should be encouraged for those who are eligible.To reduce gathering, maintain social distance, wear masks, quit smoking, reduce alcohol consumption, balance nutrition, and exercise properly are key tips of lifestyle intervention for COVID‐19 prevention.Active weight control with a target body mass index (BMI) < 25 kg/m(1)Glycated hemoglobin (HbA1c) control target: In general, HbA1c control target is <7%. Appropriate adjustment in combination with patient's age, diabetes duration, comorbidities, risk of hypoglycemia, and expected life span is necessary.(2)For diabetes patients with hypertension, the blood pressure control target is generally <130/80 mm Hg.(3)Low‐density lipoprotein cholesterol should be controlled below 2.6 mmol/L (100 mg/dl) for diabetes patients and further adjustment according to the patient's atherosclerotic cardiovascular disease risk is appropriate.(4)Try to use medications with sufficient clinical evidence including metformin, glucagon‐like peptide‐1 receptor agonists and sodium‐glucose cotransporter 2 inhibitors for glycemic control, statins for lipid lowering, and renin‐angiotensin system inhibitors for blood pressure control, so as to reduce the risks for cardiovascular disease comorbidities, diabetic microvascular complications, and all‐cause mortality in type 2 diabetes patients.Comprehensive management of blood glucose, blood pressure, and lipid levels according to the principles of hierarchical management should be achieved.Adequate rest and nutritional support should be strengthened; vital signs, biochemical parameters, and chest imaging should be monitored; at the same time, antiviral, supportive treatment and other comprehensive management could be performed. Details can be found in the “Diagnosis and treatment plan for COVID‐19 ”.1Basic requirements(1)Ensuring adequate sleep, setting up a regular routine, and self‐monitoring of blood glucose are all necessary.(2)Reduce the intake of foods with high glycemic index, such as starches, high‐sugar fruits, sugar‐sweetened beverages, etc., and appropriately increase the intake of high‐quality protein.(3)Taking moderate or low‐intensity exercise, about 150 min a week, divided into 4‐5 times to accomplish the sports. Selecting an appropriate exercise mode according to the patient's physical condition and range of motion are necessary. Marching on the spot, tai chi, and calisthenics are optional indoor sports during the pandemic.2(1)Patients aged <65 years and without obvious chronic complications of diabetes: maintaining fasting blood glucose at 6.1–7.8 mmol/L and 2‐h postprandial blood glucose: 7.8–10.0 mmol/L is recommended.(2)Patients aged ≥65 years, or with serious diabetes related comorbidities, a less stringent fasting blood glucose at 7.8–10.0 mmol/L, 2‐h postprandial blood glucose at 7.8–12.0 mmol/L is recommended.(3)Previous medications can be maintained for those with stable glycemic control status. If blood glucose concentration increases continuously, switching to insulin administration is preferred.(4)If hypoglycemic symptoms  occur, fingerstick blood glucose should be taken immediately, and sugary foods should be taken as soon as possible, such as juice, glucose water, candy, etc. If the symptoms cannot be relieved or aggravated, an intravenous bolus of glucose should be given quickly. At the same time, discontinuing current hypoglycemic agents and monitoring blood glucose are necessary. Hypoglycemic medication should be adjusted subsequently.Target of glycemic control and medication selection for asymptomatic/mild symptoms patients3For patients hospitalized with common or severe COVID‐19: stratified management of blood glucose and subcutaneous insulin administration or intravenous insulin injection for certain conditions is recommended.With the continuous development of Internet of things technology, telemedicine has been applied widely. Therefore, during the COVID‐19 pandemic, the use of Internet‐based platforms outside of hospitals including specialized medical apps or certain social software are all efficient supplements for remote consultation and monitoring for sustained metabolic control and medication adjustment for these patients, as well as reducing the potential risks associated with hospital visits.None declared.


Transthyretin‐related cardiac amyloidosis (TTR‐CA) is thought to be particularly common in specific at‐risk conditions, including aortic stenosis (AS), heart failure with preserved ejection fraction (HFpEF), carpal tunnel syndrome (CTS) and left ventricular hypertrophy or hypertrophic cardiomyopathy (LVH/HCM).We performed a systematic revision of the literature, including only prospective studies performing TTR‐CA screening with bone scintigraphy in the above‐mentioned conditions. Assessment of other forms of CA was also evaluated. For selected items, pooled estimates of proportions or means were obtained using a meta‐analytic approach.p = .04), LVH/HCM  and CTS (82 vs. 71), but not in HFpEF . In terms of comorbidities, hypertension, diabetes and atrial fibrillation were highly prevalent in TTR‐CA‐diagnosed patients, as well as in those with an implanted pacemaker.Nine studies  accounting for 1375 screened patients were included. One hundred fifty‐six (11.3%) TTR‐CA patients were identified . Exclusion of other forms of CA and use of genetic testing was overall puzzled. Age at TTR‐CA recognition was significantly older than that of the overall screened population in AS (86 vs. 83 years, Screening with bone scintigraphy found an 11–15% TTR‐CA prevalence in patients with AS, HFpEF and LVH/HCM. AS and HFpEF patients were typically older than 80 years at TTR‐CA diagnosis and frequently accompanied by comorbidities. Several studies showed limitations in the application of recommended TTR‐CA diagnostic algorithm, which should be addressed in future prospective studies. The analysis was conducted using R .We performed a systematic revision of the literature in PubMed/Embase to identify prospective studies conducting screening for TTR‐CA by the means of bone scintigraphy in each of the following at‐risk conditions: AS undergoing replacement; HFpEF; CTS; and LVH/HCM. The search is updated to December 2020. Reviews and retrospective studies, as well as those performing screening by means of genetics or cardiac magnetic resonance or endomyocardial biopsy or intraoperative biopsy or autopsy—without prior or concomitant bone scintigraphy—were excluded Figure . Moreove33.1Nine studies, accounting for 1375 screened patients, were included in the present analysis: 3 for AS,99mTc‐3,3‐diphosphono‐1,2‐propanodicarboxylic acid (99mTc‐DPD) was the scintigraphic bone tracer more commonly used (in 5 studies). Two AS studies3.2Overall, 156 (11.3%) TTR‐CA patients were identified by means of screening with bone scintigraphy. The prevalence of TTR‐CA varied between settings: 11.4% in AS, 14.8% in HFpEF, 2.6% in CTS and 12.9% in LVH/HCM. Out of the 156 TTR‐CA identified patients, 140 (89.7%) were TTRwt‐CA and 16 (10.3%) were TTRv‐CA. All TTRv‐CA patients were identified in the LVH/HCM setting. Recognition of other CA aetiologies was low, with a total of 31 (2.3%) patients, mostly identified in the LVH/HCM setting (23 out of 31 patients).p = .12), but significantly greater in AS than in CTS (71 years) or LVH/HCM . In AS studies, there were no inclusion/exclusion age cut‐offs, which were instead set in both HFpEF studies (≥60 and 65 years),p = .04), LVH/HCM  and CTS , but not in HFpEF .Table p < .01 for all). The presence of arterial hypertension was comparable in patients with and without TTR‐CA within each condition. About one fourth of AS and HFpEF screened patients had diabetes, with a similar rate in those with and without TTR‐CA. Atrial fibrillation appeared relatively more common, within each condition, in patients with TTR‐CA than in the screened population , and significantly more frequent in TTR‐CA HFpEF patients than in TTR‐CA associated with other conditions . Finally, about 10% of screened patients within each condition had a pacemaker already implanted; this percentage doubled or tripled in patients diagnosed with TTR‐CA , but not in those with CTS (4.5 vs. 0.0%).In terms of comorbidities, AS and HFpEF screened populations were more frequently affected by arterial hypertension as compared to CTS and LVH/HCM patients (about 81.6% vs. about 61.1%, respectively, Outcomes were assessed only in AS studies, which reported no differences in terms of mortality between AS patients with and without CA.4We herein present a systematic revision of prospective studies that performed bone scintigraphy screening for TTR‐CA in different populations at risk. Our main findings pertain the methodology of these studies and the epidemiological scenario that they depict.4.1Although conditions deemed at risk of TTR‐CA have been long and widely identified,The two HFpEF and CTS studies had different inclusion criteria and may have not been completely comparable: Bennani Smires and colleaguesFinally, genetic TTR testing was not routinely performed in two AS study,4.2Apart from CTS, in which TTR‐CA prevalence was 2.6%, the overall yield of screening for TTR‐CA ranged between 11.4% and 14.8%. However, characteristics of screened and diagnosed patients varied significantly across different conditions (Table Increasing the awareness towards TTR‐CA remains of critical importance in everyday clinical practice and in clinical trials.5Studies screening for TTR‐CA in specific populations considered at risk found a varying prevalence, particularly relevant in AS, HFpEF and LVH/HCM. Nevertheless, characteristics of TTR‐CA patients diagnosed in these settings were different, and those with AS and HFpEF were typically older than 80 years of age and with multiple comorbidities. Considering costs and efficacy of contemporary available disease‐modifying therapies, these results question whether these conditions represent adequate screening settings for the early recognition of TTR‐CA. Moreover, several of the screening studies included in the present work have inherent methodological limitations that may have partly hindered their accuracy and that should be addressed in future prospective studies.All authors reported no conflicts related to the present work. M.C. received speaker and/or advisor fees in the last 2 years from Akcea Therapeutics, Menarini, Novartis, Pfizer, Sanofi e Sanofi Genzyme, Vifor Pharma and two investigator‐initiated grants from Pfizer, all outside of the scope of the present work.Supplementary MaterialClick here for additional data file.


Liver transplantation (LT), resection (LR), and ablation (LA) are three curative‐intent treatment options for patients with early hepatocellular carcinoma (HCC). We aimed to develop a prognostic calculator to compare the long‐term outcomes following each of these therapies.A total of 976 patients with HCC within the Milan criteria who underwent LT, LR, and LA between 2009 and 2019 from four institutions were evaluated. Multistate competing risks prediction models for recurrence‐free survival (RFS), recurrence within the Milan criteria (RWM), and HCC‐specific survival (HSS) were derived to develop a prognostic calculator.p < 0.001). For multinodular HCC ≤3 cm, although LT had better RFS and HSS than LR or LA, LA was noninferior to LR. An online prognostic calculator was then developed based on the preoperative clinical factors that were independently associated with outcomes to evaluate RFS, RWM, and HSS at different time intervals for all three treatment options.During a median follow‐up of 51 months, 420 (43%) patients developed recurrence. In the multivariate analysis, larger tumor size, multinodularity, older age, male, higher alpha‐fetoprotein (AFP), higher albumin‐bilirubin (ALBI) grade, and the presence of portal hypertension were significantly associated with higher recurrence and decreased survival rates. The RFS and HSS were both significantly higher among patients treated by LT than by LR or LA and significantly higher between patients treated by LR than by LA  within the Milan Criteria (MC), three curative‐intent treatment options are liver transplantation (LT), resection (LR) and ablation (LA). This online prognostic calculator was developed based on the preoperative clinical factors that were independently associated with outcomes to evaluate and compare recurrence‐free survival (RFS), recurrence within the Milan criteria (RWM), and HCC‐specific survival (HSS) at different time intervals for all three treatment options. Patients who recurred within the MC are potentially eligible for salvage transplantation. This would help physicians guide patients in a personalized and informed discussion of their treatment options.22.1createDataPartition” in the caret package to do this. Secondly, the remaining samples consisted of the validation cohort. Lastly, we further examined the balance between the training and validation cohorts; the details are shown in Table With the approval of the institutional review board at each center, we retrospectively identified 976 patients between 2009 and 2019 who had initial presentation of treatment‐naïve HCC within the MC at four medical centers. The Chinese cohort consisted of 716 patients who were then randomly assigned to the Chinese training cohort and the Chinese validation cohort in a 2:1 ratio to develop the three prognostic models for RFS, RWM, and HSS. These three models were then externally validated with the American cohort which consisted of 260 patients. For the Chinese cohort, we applied the simple random sampling method in the following steps. Firstly, we drew a random sample from the Chinese cohort to be in the training cohort by simple random sampling without replacement, for which each patient has equal probability of being selected. The sample size of this training cohort was approximately two‐third of the whole cohort. We used R function “The study included adult patients who fulfilled all of the following inclusion criteria: (i) new diagnosis of HCC based on either imaging or histological evaluation according to the European Association for the Study of the Liver (EASL) guidelinesPatients were excluded if they had prior treatment for HCC, had Child–Pugh class C cirrhosis, or underwent LT, LR, or LA with non‐curative intent. Thus, patients who underwent LA while having additional untreated tumors as an attempt to bridge to LT were excluded. Patients who had concomitant LR, LA, and/or transcatheter arterial chemoembolization (TACE) were also excluded. All patients in this study underwent either LT, LR, or LA, which included either radiofrequency or microwave ablation.In addition to comparing the long‐term outcomes of the three treatment groups  using RFS, RWM, and HSS, three additional subgroups under each treatment group were also further evaluated according to their tumor number and size: (i) solitary tumor ≤3 cm; (ii) multifocal tumors (2–3 tumors), each ≤3 cm; or (iii) a solitary tumor 3–5 cm.9/L. Patients' actual treatments were decided generally based on the EASL guidelinesThe severity of liver disease was evaluated using the albumin‐bilirubin (ALBI) grade, Child–Pugh grade, aspartate aminotransferase‐to‐platelet ratio index (APRI) score, and Model for End‐Stage Liver Disease (MELD) score. Portal hypertension was diagnosed with indirect measures of ascites, and/or presence of gastroesophageal varices, and/or splenomegaly with a platelet count of less than 100 × 102.2Following LT, LR, or LA, patients were evaluated using α‐fetoprotein (AFP), liver function tests, and computed tomography during follow‐up every 3 months during the first 2 years and every 6 months thereafter. Recurrences were classified as within the MC or beyond the MC at the time of the first recurrence.The primary endpoint of our study was to compare RWM, RFS, and HSS among patients who underwent LT, LR, or LA. RWM was the time from the date of treatment to the date of the first recurrence within the MC. Patients who recurred beyond the MC or were recurrence‐free at the date they were last known to be alive were censored at this time. RFS was defined as the time from treatment to recurrence or death, whichever was observed first. Patients were censored if they were alive without recurrence at the last visit. HSS was defined as the time of diagnosis to recurrence‐related death, and patients were censored if they were alive at the last visit or died from non‐HCC‐related causes.2.3Continuous variables were presented as medians and interquartile ranges, whereas categorical variables were presented as frequencies and proportions. Continuous variables in the training and validation cohorts were compared using Kruskal–Wallis tests. Categorical variables were compared using chi‐square tests or Fisher's exact tests, as appropriate. Nonparametric analyses of RFS and HSS for the three treatment groups were performed with Kaplan–Meier (K–M) estimators and compared with log‐rank tests. Competing risks analyses were conducted to estimate the cumulative incidence functions of the RWM and test equality across groups. In order to overcome the treatment selection bias in the Chinese training cohort, we performed the inverse probability of treatment weighting (IPTW) based on the method of the propensity score to evaluate the three options.Univariate Cox PH models of RFS, RWM, and HSS were used to examine the association of the responses with each covariate separately. Since the three treatments were found to have different effects in the solitary and multiple HCC patients in our study, we developed multivariate Cox PH models including the interaction term between treatment allocation and tumor number. Stepwise regression was used to select the final model with the smallest Akaike information criterion (AIC). To model and predict the recurrence within the MC, we conducted a competing risks model and developed a multivariate cause‐specific Cox PH model based on stepwise selection, treating recurrence beyond the MC, and death for other than HCC as competing risks. To analyze the HSS, we performed another competing risk model for death due to HCC, treating non‐HCC‐related death as a competing risk.p < 0.05 was used to indicate statistical significance and all statistical tests were two sided.To evaluate and validate model performance, discrimination abilities and calibration were evaluated for the Chinese training cohort, Chinese validation cohort, and American validation cohort. A receiver operating characteristic (ROC) curve and the concordance index were used to assess the discrimination, and a calibration plot comparing the predicted probability with the observed probability was used to assess the calibration ability. The risk scores were used to stratify patients into low‐, intermediate‐, and high‐risk groups. An online prognostic calculator to predict the RFS, the RWM, and the HSS was developed as an interactive web application using R Shiny App.33.1n = 311), LR (n = 312), or LA (n = 353) as their first‐line treatment , LR (n = 247), or LA (n = 226) as their first‐line treatments were randomly assigned to the Chinese training cohort (n = 482) and Chinese validation cohort (n = 234) in a 2:1 ratio. The long‐term outcomes of the training and validation cohorts were matched well in that the RFS, RWM, and HSS were not significantly different between them  , LR (n = 65), or LA (n = 127). The clinicopathologic characteristics of the Chinese training cohort (n = 482), the Chinese validation cohort (n = 234), and the American validation cohort (n = 260) were summarized , 420 (43%) patients developed recurrence, of whom 245 (25.1%) experienced recurrence within MC, and 165 (16.9%) patients experienced recurrence beyond MC. Of 157 (44.5%) patients in the LA groups, only four patients underwent salvage transplantation. In contrast, only 10 (3.2%) patients in the LT group developed recurrence with the MC.Among 243 transplanted patients in the Chinese training and validation cohorts, the median time spent on the waiting list in the LT group was 3.7 months, and 187 (77.0%) patients underwent bridging therapy prior to LT. This is in contrast to the American validation cohort, in which the median waitlist time was 8.4 months, and 67 (99%) underwent at least one bridging therapy prior to LT. This did not account for additional patients who had dropped out of the long waitlist for LT due to cancer progression or death.In total, 309 (31.7%) patients died after a median follow‐up of 50.8 months, of which 244 (25%) patients' deaths were due to recurrence, and 65 (6.7%) patients' deaths were due to non‐HCC causes, which included graft malfunction, chronic rejection, acute liver failure, comorbidities, or other complications.3.3Within the Chinese training cohort, clinicopathological variables among the LT, LR, and LA groups showed significant differences Table . After tIn the Chinese training group, univariate analyses were performed to evaluate clinicopathological factors significantly associated with RFS, RWM, and HSS Table . The mulIn the subgroup analysis of tumors by size and multinodularity, for solitary tumors ≤3 cm or 3–5 cm, LT had significantly better RFS and HSS than LR or LA, and LR had significantly better RFS and HSS than LA Table . However3.4The RFS, RWM, and HSS among LT, LR, and LA treatment groups showed similar results in the entire cohort as in the Chinese training cohort, the Chinese validation cohort, and the American validation cohort Figure .p < 0.001) , which were likely due to smaller sample size.In the entire cohort, the RFS and HSS were both significantly longer in patients treated by LT than LR or LA, and significantly longer in patients treated by LR than LA  Figure . This wap < 0.001)  Figure . This fiThe risk scores based on the final multivariate models in the Chinese training cohort individually stratified the patients into three prognostic strata corresponding to low, intermediate, and high risks. Using −0.7 and 0.8, 0.2 and 1.1, −0.8 and 1 as the cutoff values for the RFS prediction model, the RWM prediction model, and the HSS prediction model, respectively. The low‐, intermediate‐, and high‐risk probabilities of RFS, RWM, and HSS based on the three models were well stratified among the entire cohort, Chinese training cohort, Chinese validation cohort, and American validation cohort Figure .3.5The summary statistics c‐indexes of 2‐year RFS and RWM, 3‐year RFS, RWM, and HSS, and 5‐year RFS and HSS in the Chinese training cohort, Chinese validation cohort, and American validation cohort demonstrated the satisfactory predictive performance of these RFS, RWM, and HSS models Table . The calhttps://yingwu.shinyapps.io/Tianjin‐PC‐MDT/. By inputting preoperative clinical factors such as tumor size and number, gender, age, AFP level, ALBI grade, and portal hypertension status, the online calculator can reveal RFS, RWM, and HSS at different time intervals for all three treatment options. For example, in a 50‐year‐old male patient with a solitary HCC ≤3 cm, AFP 10 ng/ml, ALBI grade 1, and absence of portal hypertension, RFS, and HSS were higher when treated by LT than LR, and by LR than LA .In the subgroup analysis, while LR had significantly better RFS and HSS than LA for solitary tumors ≤3 cm or 3–5 cm, the LA approach was noninferior to LR for multinodular tumors . One potential explanation is that multinodularity is a marker of an aggressive tumor biology, as it is associated with microvascular invasion, poor differentiation, and liver cirrhosis at risk for multiple de novo development of HCC or recurrence at the treated site. Because of this, the risk of recurrence would be high regardless of local LA or oncologic LR.Microvascular invasion is a well‐known risk factor for recurrence following treatment of HCC.Several prognostic systems have been developed to stratify patients with HCC, including the Barcelona Clinic Liver Cancer (BCLC) staging,While other research groups have evaluated the prognosis of patients with HCC, they typically focused on comparing one or two treatment modalities instead of evaluating all three curative options. To our knowledge, given scarce organ limitations of LT and ethical implications involved, there is no randomized controlled trial exists comparing all three modalities.N = 8, 3.1%), hepatitis B + C , hepatitis C , alcohol use , and non‐alcoholic steatohepatitis . Thus, further studies especially multi‐institutional, prospective studies involving the global population would be required to further validate our model.There are several limitations in this study. First, because of its retrospective and non‐randomized design, the baseline characteristics of the three treatment groups in the Chinese training cohort were different and thus may cause inherent biases when developing the prediction models. However, we did use IPTW adjustment to overcome these biases and got similar characteristics before and after the IPTW adjustments which improved the reliability of the prediction models. Second, although antiviral treatment was generally utilized for hepatitis B/C‐related HCC prior to these curative treatments, it was not included in our models because it was not documented in our databases. Third, because the majority of Chinese patients with HCC were related to hepatitis B cirrhosis , they may present different tumor behaviors from the American patients with HCC as their cirrhosis was due to several other etiologies.5Preoperative clinical factors including tumor size and number, gender, age, AFP level, ALBI grade, and portal hypertension status were independently associated with recurrence‐free survival, recurrence within the Milan criteria, and HCC‐specific survival following liver transplantation, resection, or ablation. A prognostic online calculator was then created using these preoperative clinical factors to compare long‐term outcomes following these three first‐line treatment options. LT had better outcomes than LR and LA among all comers, but its utilization was limited by scarce organ availability. While the outcomes of LR were better than LA in patients with solitary HCC, their outcomes were comparable in patients with multinodular HCC. This online calculator is clinically useful, easy to use in everyday practice, and readily available online to guide clinicians in informed discussion of the optimal treatment strategies with patients and during multidisciplinary tumor board.Concept and design: NNZ, JZ, DG, WL, JHH, YMZ, WTJ, TQS. Data collection: JZ, VK, JYL, JG, YHS, SWZ, TL, SW, WL, YBQ. Provision of study materials or patients: ST, ZYS, WZ, LZ, YX, YHZ, JYL. Statistical analysis: YW, NNZ, JZ, YMZ, WTJ, TQS, WL, JHH, DG. Writing of article: All authors.This work was supported by a grant from the Key Project of Science and Technology, Tianjin Science and Technology Committee (19YFZCSY00020).The authors report no conflicts of interest in this study.This study was conducted with the approval of the University of Pittsburgh (STUDY20090034), Tianjin Medical University Cancer Institute (bc2019082) and Hospital Medical Ethics Committee, Tianjin First Center Hospital Medical Ethics Committee (2016N070KY), and Tianjin Second People's Hospital Medical Ethics Committee, Tianjin, China (2016N043X).Appendix S1Click here for additional data file.


Eventually, postoperative pathological diagnosis demonstrated TTH. So far, there are no reports of 18F‐FDG PET/CT imaging of this disease.True thymic hyperplasia (TTH) in children is rare and difficult to distinguish from other thymic tumors such as thymoma and thymic carcinoma. A 3‐year‐old girl underwent an  The enlarcinoma . However2, firm in texture, with poor mobility, and with no sternal pressure and rubbing sensation in the chest. She had no other symptoms and no family history of TTH or other diseases. Laboratory tests showed a mildly elevated lactate dehydrogenase (LDH) at 276 U/L (reference range: 38–126 U/L), a white blood cell count of 13.76 × 109/L (reference range: 4–10 × 109/L), and a red blood cell count of 5.09 × 1012/L (reference range: 4–4.5 × 1012/L). However, thyroid‐related laboratory test results were normal. In addition, neuron‐specific enolase (NSE) and human chorionic gonadotropin levels (HCG) were normal. The chest CT scan revealed a well‐defined irregular mass of 17 × 10 × 7.5 cm3 in the right anterior mediastinum, with strip‐like low‐density shadows in the lesion, and a CT value of approximately −67 Hounsfield units, thereby suggesting adipose tissue  . TTH is TTH in children is a rare clinical condition, has unknown etiology, and often has no comorbidities. It is often referred to as massive thymic hyperplasia (MTH) because it is characterized by a relatively larger and heavier thymus than the thymus of a healthy individual. The diagnostic features of MTH are as follows: 1. radiographs show gland projection exceeding the cardiac shadow, 2. The thymus weighs several times its expected weight at a specific age, 3. the mass of the thymus is > 2% of the body mass, and 4. pathological test shows normal thymus structure . The pat18F‐FDG PET/CT can differentiate between benign and malignant thymic tumors but cannot distinguish between aggressive and non‐aggressive thymoma and still relies on morphological examination by CT and MRI or pathological examination ’ legal guardian/next of kin for the publication of any potentially identifiable images or data included in this article.JR acquisition of data, drafting of the manuscript; ZF and YZ: revision of the manuscript, supervision. All authors read and critically revised the manuscript for intellectual content and approved the final manuscript.