No
Yes (SSN not exposed)
Yes (SSN exposed)
Table 2 presents the prevalence of identity theft victimization overall and by subtype. The prevalence of overall identity theft victimization (any type) was 6.2% in the combined 2012/2014 sample (95%CI = 6.0%–6.3%). The most common form of victimization was existing credit card or bank account identity theft, with a prevalence of 5.6% (95%CI = 5.5%–5.8%).
Identity theft victimization frequencies.
Identity Theft Victimization Subtype | Combined 2012/2014 (n = 128,419) |
---|---|
n (%) | |
Any subtype | 7921 (6.2) |
Existing credit or bank account | 7241 (5.6) |
New accounts | 492 (0.4) |
Instrumental purposes | 350 (0.3) |
Table 3 presents results from the multivariable analysis of risk and protective factors of identity theft victimization for each subtype. Higher levels of online purchasing behavior were significantly associated with increasing odds of existing credit card/bank account and new accounts identity theft victimization; those engaging in daily online shopping were more than five times as likely to be victims of existing credit card/bank account identity theft as those not engaging in online purchasing (OR = 5.74, 95%CI = 4.31–7.64). Persons reporting breached personal information from a company or government were significantly more likely to experience identity theft, particularly if social security information was exposed (instrumental purposes: OR = 8.05, 95%CI = 5.66–11.46; new accounts: OR = 3.83, 95%CI = 2.67–5.51; existing credit/bank account: OR = 1.46, 95%CI = 1.26–1.68). Those reporting other NCVS victimizations were between 29% (existing credit/bank account: OR = 1.29, 95%CI = 1.23–1.35) and 46% (new accounts: OR = 1.46, 95%CI = 1.32–1.62) more likely to be victims of identity theft with each successive crime. Individuals with a history of identity theft victimization were 28% more likely to be victimized by existing credit/bank account identity theft in the past year than those with no prior history (OR = 1.28, 95%CI = 1.19–1.37).
Multivariable logistic regression models predicting identity theft victimization.
Independent Variables | Existing Credit or Bank Account (n = 116,042) | New Accounts (n = 128,419) | Instrumental (n = 128,419) |
---|---|---|---|
OR (95% CI) | OR (95% CI) | OR (95% CI) | |
Online purchasing behavior frequency (ref. None) | |||
Up to once per month (1–12 times/year) | 2.45 (2.28–2.63)*** | 1.71 (1.35–2.17)*** | 1.35 (1.02–1.78) |
Up to once per week (13–52 times/year) | 3.54 (3.27–3.83)*** | 1.78 (1.33–2.38)*** | 1.12 (0.77–1.64) |
Up to once per day (58–365 times/year) | 4.44 (4.02–4.90)*** | 1.89 (1.25–2.85) | 2.01 (1.28–3.16) |
More than once per day (More than 365 times/year) | 5.74 (4.31–7.64)*** | 4.52 (1.79–11.46) | 4.03 (1.39–11.70) |
Number of other victimizations (cont.) | 1.29 (1.23–1.35)*** | 1.46 (1.32–1.62)*** | 1.41 (1.24–1.60)*** |
Breached personal information (ref. No) | |||
Yes (SSN not exposed) | 1.44 (1.33–1.56)*** | 1.96 (1.44–2.66)*** | 2.16 (1.47–3.19)*** |
Yes (SSN exposed) | 1.46 (1.26–1.68)*** | 3.83 (2.67–5.51)*** | 8.05 (5.66–11.46)*** |
Identity theft victimization prior to past year (ref. No) Yes | 1.28 (1.19–1.37)*** | 1.43 (1.11–1.85) | 1.43 (1.05–1.95) |
Purchase protective services (cont.) | 1.02 (0.95–1.09) | 1.62 (1.28–2.06)*** | 1.37 (0.99–1.87) |
Routine protective behaviors (cont.) | 0.76 (0.75–0.78)*** | 0.66 (0.61–0.71)*** | 0.71 (0.65–0.78)*** |
Age generations (ref. millennials) | |||
Generation X | 1.21 (1.12–1.29)*** | 1.28 (1.00–1.65) | 1.68 (1.26–2.24)*** |
Baby boomers | 1.38 (1.29–1.48)*** | 1.70 (1.32–2.20)*** | 1.79 (1.32–2.42)*** |
Silent or Greatest | 1.10 (0.99–1.21) | 1.23 (0.86–1.78) | 1.12 (0.72–1.75) |
Gender (ref. Male) Female | 0.99 (0.94–1.04) | 0.95 (0.79–1.13) | 1.14 (0.92–1.42) |
Marital Status (ref. Married/partnered) Not married/partnered | 0.95 (0.90–1.01) | 1.23 (1.00–1.51) | 1.63 (1.28–2.09)*** |
Educational attainment (ref. High school or less) | |||
Some college or associate degree | 1.42 (1.33–1.52)*** | 1.70 (1.35–2.14)*** | 1.43 (1.11–1.86) |
Bachelor’s degree | 1.67 (1.56–1.80)*** | 1.66 (1.25–2.20)*** | 1.18 (0.84–1.66) |
Graduate/professional degree | 1.90 (1.74–2.07)*** | 1.85 (1.31–2.61) | 0.95 (0.59–1.50) |
Race/ethnicity (ref. non-Hispanic white) | |||
Hispanic | 0.85 (0.78–0.93)*** | 1.32 (1.00–1.73) | 0.93 (0.66–1.32) |
Black | 0.78 (0.71–0.86)*** | 1.43 (1.11–1.86) | 1.58 (1.20–2.09) |
AAPI/AIAN | 0.78 (0.70–0.87)*** | 0.73 (0.46–1.16) | 0.69 (0.39–1.22) |
Other | 1.09 (0.89–1.32) | 3.32 (2.17–5.09)*** | 1.18 (056–2.50) |
Household income (ref. $0 to 24,999) | |||
$25,000 to 49,999 | 1.05 (0.95–1.15) | 0.77 (0.60–1.00) | 0.90 (0.67–1.21) |
$50,000 to 74,999 | 1.20 (1.08–1.33) | 0.73 (0.54–0.99) | 0.80 (0.56–1.13) |
$75,000+ | 1.38 (1.25–1.52)*** | 0.71 (0.52–0.97) | 0.74 (0.52–1.05) |
Number of household members ≤ 12 years (cont.) | 1.01 (0.98–1.05) | 1.20 (1.08–1.33) | 1.21 (1.07–1.36) |
Residential setting (ref. urban) Rural | 0.90 (0.84–0.96) | 0.80 (0.61–1.05) | 0.65 (0.46–0.91) |
Interview type (ref. In-person) Telephone | 0.91 (0.87–0.96)*** | 0.85 (0.71–1.02) | 0.74 (0.60–0.92) |
Note: All multivariable logistic regression models, except the New Accounts model, satisfied the Omnibus Test of Model Coefficients (p < 0.01). All multivariable logistic regression models satisfied the Hosmer-Lemeshow Test (p > 0.05). Across models, independent variables had tolerance of 0.70 or above and variance inflation factor of 1.43 or below, indicating no concern of multicollinearity.
CI = Confidence interval; OR: Odds ratio; SSN: Social Security Number; AAPI/AIAN = Asian American/Pacific Islander/American Indian/Alaskan Native. ***p < 0.001, (two-tailed tests).
Individuals engaging in a higher number of proactive, routine protective behaviors, such as shredding documents and updating passwords, were between 25% (existing credit/bank account: OR = 0.76, 95%CI = 0.75–0.78) and 35% (new accounts: OR = 0.66, 95%CI = 0.61–0.71) less likely to experience identity theft victimization with each additional protective behavior. Purchasing credit monitoring services and identity theft insurance, however, was associated with significantly higher odds of new accounts (OR = 1.62, 95%CI = 1.28–2.06) identity theft.
Across all identity theft subtypes, baby boomers were most likely to be victims (existing credit/bank account: OR = 1.38, 95%CI = 1.29–1.48; new accounts: OR = 1.70, 95%CI = 1.32–2.20; instrumental: OR = 1.79, 95%CI = 1.32–2.42). Unmarried/un-partnered persons were 63% (OR = 1.63, 95%CI = 1.28–2.09) more likely to experience instrumental forms of identity theft. Higher levels of education were associated with increasingly higher odds of both existing credit card/bank account and new accounts forms of identity theft. Compared to non-Hispanic whites, existing credit/bank account victimization was less likely among Hispanic (OR = 0.85, 95%CI = 0.78–0.93), Black (OR = 0.78, 95%CI = 0.71–0.86), and AAPI/AIAN (OR = 0.78, 95%CI = 0.70–0.87) persons. Persons living in households in the highest income bracket were most likely to experience existing credit/bank account identity theft (OR = 1.38, 95%CI = 1.25–1.52) compared to those in the lowest income households. As a methodological finding, respondents who participated in a telephone rather than in-person interview were significantly less likely to report identity theft victimization.
Approximately 1 out of every 15 adults aged sixteen years or older in the U.S. – over 16 million people – experience some form of identity theft each year. In addition to direct losses, consequences may include damaged credit, legal fees, loss of trust, and health outcomes such as stress, anxiety, and depression ( Harrell, 2015 , Golladay and Holtfreter, 2017 ). Among victims who experienced the misuse of personal information for instrumental purposes, approximately 56% suffered moderate to severe distress, a similar percentage as seen among victims of violence ( Harrell, 2015 ).
As large-scale data breaches have become an unfortunate part of our growing tech-based marketplace, this analysis examined whether online purchasing behavior and personal data security practices affect the risk of identity theft victimization, or whether becoming a victim is largely contingent on corporate and government-level data breaches. Findings provide support for the L-RAT model of victimization which suggests that individual lifestyle routines and degree of protective measures/guardianship influence the likelihood of victimization.
Respondents who stated that their information was part of a large data breach were significantly more likely to report all forms of identity theft, particularly when their social security numbers were exposed. Victims of identity theft for instrumental purposes were eight times as likely to say their social security numbers were exposed in a data breach compared to non-victims, likely because that form of identity theft requires social security numbers to access government benefits and other services. Although it is not possible to assess whether data breaches directly caused identity theft incidents, data breaches were significantly correlated with the misuse of identity information.
L-RAT proposes that routine lifestyle behaviors contribute to crime victimization risk. In the present study, individual risk and protective behaviors were consistent and strong (magnitude) predictors. Similar to findings using a Canadian sample ( Reyns & Henson, 2016 ), increasing levels of online purchasing activity were associated with incrementally higher odds of financial account and new account identity theft. Participating in commercial activities online reflects a major societal innovation and lifestyle shift that has allowed consumers to purchase products conveniently and globally, but entering personal data online entrusts vendors to safely store and manage this data. For example, Holtfreter et al. (2015) found that individuals who placed an order with a company they had never done business with before were significantly more likely to be victims of identity theft. While the NCVS ITS does not ask respondents what online retailers they have made purchases from, it is likely that as the frequency of online shopping increases, the odds of using an unsecured payment portal or having information exposed in a retail data breach increases. Further innovations in online security and payment systems are required to protect users’ information, and future research should explore precisely how online purchasing activities expose personal information.
In support of the guardianship principle of L-RAT, proactive individual behaviors, like shredding personal documents and routinely changing account passwords, significantly reduced the likelihood of identity theft. Unfortunately, the Pew Research Center ( Olmstead & Smith, 2017 ) found that half of U.S. respondents were not educated about everyday security practices. Given that routine safety behaviors reduce risk of identity theft, consumer protection efforts need to focus on educating consumers on the basics of online security. Purchasing external credit monitoring and identity theft protection services did not reduce risk and was related to greater likelihood of new accounts identity theft victimization. Perhaps respondents who purchased these services had some knowledge that their identity may be misused. Another explanation is that some criminal entities have reached a level of sophistication to evolve techniques ahead of current industry protection standards ( Moore et al., 2009 ).
This study found that exposure to other types of crime, as well as prior experiences with identity theft, were associated with a greater risk of identity theft victimization. Personal information may be stolen during the course of other crimes directly (e.g., theft of wallets, bank statements) or indirectly through theft of devices that contain personal information. This result is consistent with financial fraud research—prior fraud victimization increases the odds of re-victimization ( Titus et al., 1995 ). An underground system exists for identity theft where specified pieces of stolen identifying information are bundled and sold to other criminals, thereby increasing the odds that it is used for various identity crimes over time ( Moore et al., 2009 ). Services for identity theft victims should include help contacting the major credit bureaus to place a temporary freeze or fraud alert on credit reports to prevent criminals from opening new accounts with victims’ stolen credentials.
The socioeconomic and demographic risk patterns found in this study were roughly consistent with the predictions of L-RAT. In general, members of Generation X and the baby boomers, now between the ages of 39 and 73, were at the highest risk of most types of identity theft. This likely reflects the socioeconomic capacity and consumption patterns among Generation X and baby boomers relative to millennials. Together, these older generations constitute the bulk of the U.S. workforce and, therefore, have the economic means to engage in consumer activities where identities may be exposed. Longitudinal data is needed to determine whether the association between middle to late adulthood and increased risk of identity theft is indeed due to lifestyles or whether age has an independent effect.
Compared to Hispanic, Black, and Asian respondents, White respondents and those with higher educational attainment experienced significantly higher risk of existing credit card/bank account identity theft. Individuals with higher socioeconomic status have more purchasing power ( Charron-Chénier et al., 2017 ), have more access to credit ( Haushofer & Fehr, 2014 ), own more internet-enabled devices that store and transfer personal information, and are more likely to use credit cards ( Greene & Stavins, 2016 ). In support of L-RAT, this suggests that the association between existing credit card/bank account identity theft and demographic/socioeconomic profiles is related to lifestyle factors where there is greater reliance on these financial instruments, and thus more opportunities for criminals to intercept account information.
While the NCVS Identity Theft Supplement is one of the most comprehensive sources of data on identity theft, the survey likely underestimates the true extent of the problem. First, the NCVS excluded adult sub-populations who may be particularly vulnerable, such as those living with cognitive impairment and/or in institutional settings. Second, the literature on financial fraud victimization finds that people tend to under-report victimization in survey research ( Beals et al., 2015 ), and this self-report error likely extends to the issue of identity theft. Finally, the nonresponse group is likely disproportionately represented by victims who are reluctant to provide personal information in response to a survey. Another limitation of the study was that data on other potentially important behavioral variables, such as the extent of online downloading, online financial account management, types of websites visited, and presence of malware, hacking or phishing events, were unavailable. To better understand risk of identity theft victimization within the L-RAT paradigm, measures are needed to account for system-level security practices among corporate and government entities, but this is beyond the scope of the NCVS.
Identity theft victimization affects tens of millions of Americans each year. Financial exploitation, in general, is associated with major health-related consequences such as increased rates of hospitalization and all-cause mortality. Victims of identity theft experience severe mental/emotional distress, particularly among minority and older adult populations ( Harrell, 2019 , Golladay and Holtfreter, 2017 ). Given the increasing scope of this problem, the development of effective primary prevention strategies is critically needed and should focus on promoting relatively unintrusive and feasible everyday practices such as routinely changing financial account passwords, shredding documents, and checking credit reports and financial statements. The prevalence of this problem indicates that healthcare professionals will encounter patients who are victimized by identity theft on a regular basis. Healthcare settings represent an important place to both recognize vulnerable adults and provide victims with preventive education to mitigate the risk of identity exposure.
This study comprehensively examined the risk of different forms of identity theft victimization in the U.S. Although other research indicates that Americans have inadequate knowledge of cybersecurity practices ( Olmstead & Smith, 2017 ), findings from the current study demonstrated the importance of this knowledge in keeping personal information safe. Yet individual actions alone are not enough. As investment in cybersecurity grows, criminals respond with increasingly sophisticated and evolving techniques such as hacking, malware, and skimming to overcome these controls ( Pontell, 2009 ). Reducing the incidence of identity theft requires greater public/private investment in robust, dynamic data security systems and encryption tools, and more collaboration between criminal justice and law enforcement agencies to investigate and prosecute identity theft crimes.
David Burnes: Conceptualization, Formal analysis, Data curation, Writing - original draft, Writing - review & editing. Marguerite DeLiema: Conceptualization, Writing - original draft, Writing - review & editing. Lynn Langton: Conceptualization, Methodology, Writing - original draft, Writing - review & editing.
The authors declare that they have no known competing financial interests or personal relationships that could have appeared to influence the work reported in this paper.
Appendix B Supplementary data to this article can be found online at https://doi.org/10.1016/j.pmedr.2020.101058 .
Multiple Correspondence Analysis Discrimination Measures Plot.
The following are the Supplementary data to this article:
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Please note you do not have access to teaching notes, identity theft and university students: do they know, do they care.
Journal of Financial Crime
ISSN : 1359-0790
Article publication date: 30 September 2014
This study aims to explain what factors influence the relationship between the university students’ knowledge of the risk of identity theft and the preventive measures they take.
A series of semi-structured interviews was used as the primary data collection tool. The sample for this study comprised 12 undergraduate students (six males and six females) from the Flinders Business School. The interviews were designed as face-to-face interviews.
The current findings indicate that, despite the fact that students were reasonably knowledgeable regarding the general risk of identity theft, many of the students had only limited knowledge about specific issues related to identity theft. It was found that the limited knowledge or misunderstanding of specific issues prevented students from using appropriate measures that could reduce the risk of identity theft. The students demonstrated a significant misunderstanding of who perpetrators typically were targeting when stealing personal information or what perpetrators of identity theft were looking for.
The results of the study contribute to a better understanding of the students’ knowledge about the risks associated with identity crime. They may also assist governments and other stakeholders with vested interests, such as financial institutions and educational providers, to educate individuals about the circumstances where they are potentially vulnerable to identity theft.
Seda, L. (2014), "Identity theft and university students: do they know, do they care?", Journal of Financial Crime , Vol. 21 No. 4, pp. 461-483. https://doi.org/10.1108/JFC-05-2013-0032
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McALLEN, Texas – A 46-year-old resident of Mission has been taken into custody on charges of health care fraud and aggravated identity theft in connection with a scheme to defraud the Texas Medicaid Program, announced U.S. Attorney Alamdar S. Hamdani.
Juan Martin Flores, 46, is set to make his initial appearance before U.S. Magistrate Judge J. Scott Hacker at 9 a.m.
The indictment, returned Aug. 7 and unsealed upon his arrest Sept. 12, alleges he submitted or caused the submission of over 15,000 fraudulent claims to Medicaid for services that were never provided. Between 2018 and 2022, the claims resulted in approximately $2 million in Medicaid payments, according to the charges. The indictment further alleges Flores used the personal information of Medicaid beneficiaries without their consent to facilitate the fraudulent billing scheme.
According to the charges, Flores submitted claims under his national provider identifier number, representing that he provided counseling services at his office in Brownsville. However, he allegedly never actually rendered those services. The indictment details multiple instances in which he unlawfully used Medicaid beneficiaries' identities in the fraudulent claims.
Flores is charged with 10 counts of health care fraud, each carrying a possible 10-year-maximum sentence and up to a $250,000 fine. He is also facing three counts of aggravated identity theft. If convicted, he faces a mandatory two years in federal prison which must be served consecutively to any other sentence imposed.
The FBI, Department of Health and Human Services – Office of Inspector General and Texas Attorney General’s Medicaid Fraud Control Unit conducted the investigation. Assistant U.S. Attorney Andrew R. Swartz and Eric D. Flores are prosecuting the case.
An indictment is a formal accusation of criminal conduct, not evidence. A defendant is presumed innocent unless convicted through due process of law .
Two mental health care providers in the South Texas area have agreed to pay $1,083,000 to resolve False Claims Act (FCA) allegations regarding the submission of claims to Medicare, TRICARE...
Several local residents and others are now charged in six separate cases in the Southern District of Texas (SDTX) with varying counts related to the Justice Department’s 2024 National Health ...
A 68-year-old man has been sentenced for orchestrating a $6 million Medicaid fraud and kickbacks scheme
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Crime and public safety | mt. pleasant man sentenced in federal identity theft, fraud case.
United States District Judge Thomas L. Ludington on Thursday also ordered Anthony Demasi, 50, to participate in the Bureau of Prisons Inmate Financial Responsibility Program, “during which time BOP staff will develop a financial plan and assure that (Demasi) is making ‘satisfactory progress’ in meeting his financial responsibility plan.'”
Demasi, who has been free on bond since his indictment by a federal grand jury Dec. 14, 2022, was ordered to surrender “as notified by the United States Marshal,” according to the judgement document.
Demasi, who was originally charged with three counts of identity theft and three counts of fraud, owns Goldman Advisors, LLC in Mt. Pleasant, and was accused of attempting to take out credit cards from banks using the identities of people without their consent.
The federal indictment alleged that Demasi attempted to apply for credit cards from Capital One, JP Morgan Chase, and Barclay’s of Delaware in 2018, according to court records.
Bank fraud carries a maximum penalty of 30 years; identity theft is punishable by a maximum of 15 years.
Demasi entered guilty pleas to one count each of bank fraud and identity theft March 21 in front of Magistrate Judge Patricia A. Morris in Bay City.
As part of the plea agreement, Demasi is paying more than $12,000 in full restitution to two FDIC- insured banks.
Demasi previously served time in prison after pleading guilty to three counts of wire fraud and two counts of securities fraud in federal court in the Northern District of Illinois in March 2010, according to court records.
He was alleged to have lured more than two dozen victims to invest a total of roughly $4.7 million in commodity trading pools and using the money instead to fund two nightclubs in Chicago, pay gambling debts and other living expenses, and to make Ponzi-type payments to earlier investors, according to a report in the Chicago Tribune.
Mt. pleasant construction worker hit, killed on u.s. 127 in isabella county.
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npj Vaccines volume 9 , Article number: 166 ( 2024 ) Cite this article
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We compared the risks and benefits of COVID-19 vaccines using a causal pathway analysis to weigh up possible risk factors of thromboembolic events post-vaccination. The self-controlled case series (SCCS) method examined the association between thromboembolic events and vaccination while a case-control study assessed the association between thromboembolic events and COVID-19, addressing under-reported infection data issues. The net vaccine effect was estimated using results from SCCS and case-control studies. We used electronic health record data from Corewell Health (16,640 subjects in SCCS and 106,143 in case-control). We found increased risks of thromboembolic events post-vaccination (incidence rate ratio: 1.19, 95% CI: [1.08, 1.31] after the first dose; 1.22, 95% CI: [1.11, 1.34] after the second dose). Vaccination attenuated infection-associated thromboembolic risks (odds ratio: 4.65, 95% CI: [4.18, 5.17] in unvaccinated vs 2.77, 95% CI: [2.40, 3.24] in vaccinated). After accounting for vaccine efficacy and protection against infection-associated thromboembolic events, vaccination decreases thromboembolic event risk, especially during high infection rate periods.
The Coronavirus Disease 2019 (COVID-19) pandemic prompted a race to develop and distribute effective vaccines. Approximately 81.4% of the US population have been vaccinated with at least one dose, and 69.5% have completed the primary series of COVID-19 vaccination 1 . While the benefits of vaccination are widely acknowledged, concerns have emerged regarding the development of thromboembolic events after vaccination 2 . Phase 3 clinical trials were not statistically powered to identify rare adverse events 3 . The risks of new vaccines were not fully known during regulatory approval, particularly for mRNA-based vaccines (mRNA-1273 or BNT162b2), which were under authorized emergency use. Therefore, it is important to conduct post-marketing safety surveillance of the vaccines. More specifically, cases of venous thromboembolism following a mRNA-based vaccination were reported in 2022 after COVID-19 vaccines were administered in the US and some other countries 4 , 5 , 6 , 7 , drawing attention to the potential risk of thromboembolic events after the first vaccination dose. One study confirmed an increased risk of thromboembolism, ischemic stroke, and cerebral venous sinus thrombosis after the first dose of BNT162b2 8 , and another retrospective cohort study found an increased risk of cerebral venous thrombosis and portal vein thrombosis after any mRNA-based vaccination 9 . Moreover, a recent systematic review 10 has shown that thromboembolism is the most frequent cardiovascular complication following a mRNA-based vaccination. Despite those findings, vaccination is still recommended to reduce the likelihood of COVID-19, hospitalization, and mortality 8 , 11 . Furthermore, COVID-19 itself substantially increases the risk of thromboembolic events 12 , 13 , 14 , 15 , 16 , 17 , 18 , with a more prolonged and significant threat compared to vaccine-associated risks 8 . Therefore, studying the risk of thromboembolic events after COVID-19 vaccination should incorporate the protective effect of vaccines against COVID-19 severity and hence COVID-19-associated thromboembolic events.
Several studies have reported a positive correlation between thromboembolic events and mRNA-based vaccines, with reported incidence rate ratios (IRRs) between 1.04 and 1.22 8 , 19 , 20 , 21 , 22 . These studies used the self-controlled case series 23 (SCCS) design, which is a standard approach to studying adverse events of vaccines. The same design was used to evaluate the risk of thromboembolic events after COVID-19, with reported IRRs between 6.18 and 63.52 8 , 11 , 14 . However, since a thromboembolic event typically requires a hospital visit (emergency visit or hospital admission), subjects with a thromboembolic event are subject to a higher rate of COVID-19 testing, and so at a lower likelihood of misclassification as uninfected compared to subjects without an event. Hence, the SCCS design is subject to some risks of bias 24 , which we would expect to inflate the SCCS estimated relative risk (RR) of thromboembolic events after COVID-19.
The objective of this study is to evaluate whether the overall effect of the COVID-19 vaccination is to increase or decrease the risk of thromboembolic events. To do so, we first quantified the risk of thromboembolic events after mRNA-based vaccination using the SCCS method. Secondly, we evaluated the association between thromboembolic events and COVID-19 using a case-control study, avoiding the misclassification bias associated with the SCCS method. Finally, we conducted a risk-benefit analysis by comparing the magnitude of the increased risk through the direct effect of the COVID-19 vaccination with the reduced risk through the indirect pathway via protection against infection-associated thromboembolic events.
Our studies used electronic health record (EHR) data from the Corewell Health East (CHE, formerly known as Beaumont Health) and Corewell Health West (CHW, formerly known as Spectrum Health) healthcare systems, which includes demographics, mortality, hospital admissions, and COVID-19 testing. We obtained accurate COVID-19 vaccination records (vaccine types, dates, and doses) by linking EHR data at Corewell Health with the Michigan Care Improvement Registry (MCIR), giving more complete data for individuals who received the COVID-19 vaccines outside the healthcare system. We included all patients aged ≥ 18-years-old and were registered with a primary care physician within 18 months before Jan 1st, 2021.
We identified thromboembolic events based on ICD-10 (International Classification of Diseases version 10) codes from a hospital visit (emergency visit or hospital admission). These ICD-10 codes represent diagnoses for venous thromboembolism, arterial thrombosis, cerebral venous sinus thrombosis, ischemic stroke, and myocardial infarction (Supplementary Table 1 ). We also used patients with physical injury at a hospital visit (list of ICD-10 codes in Supplementary Table 2 ) to identify potential bias related to the misclassification and further leveraged them as a control group to estimate the effect of COVID-19 on thromboembolic events.
We used the SCCS design to examine the association of thromboembolic events and the first two doses of mRNA-based COVID-19 vaccines (mRNA-1273 or BNT162b2) from December 1st, 2020, to August 31st, 2022. The SCCS method compares the incidence rate of thromboembolic events before and after vaccination. In this method, subjects are under their own control, and comparisons are made within subjects, thus avoiding any time-invariant confounding. We included subjects who had a thromboembolic event and received at least one dose of the primary series of mRNA-based vaccines in the study period. The control period was defined from December 1st, 2020, to 28 days before the first dose of vaccination, excluding the period of 28 days prior to vaccination to avoid bias due to contra-indications 25 . Two separate risk periods for the first and second doses were defined until 28 days after vaccination, death, or August 31st, 2022, whichever occurred first (Supplementary Fig. 1 ). We also excluded subjects who had COVID-19 within 90 days before a thromboembolic event to remove the confounding effect of infection on that event. We used a conditional Poisson regression 22 with an offset for the length of each period to estimate the IRRs of dose one and dose two simultaneously. Specifically, the model has an independent variable of the period with three categories (control periods, and two risk periods after the first and second dose). Using the control period as the reference, we derived the IRRs for the two doses. As Poisson regression assumes the independence between recurrent events, therefore, we considered only events that occurred at least one year after the previous events.
In an initial analysis of the association between thromboembolic events and COVID-19, we used the SCCS design and included patients who had at least one positive COVID-19 test (PCR or antigen) and a thromboembolic event at a hospital visit during the same period as in the previous study of vaccination. However, due to the missing infection data in patients who did not have any hospital visits for thromboembolic events or other reasons, the SCCS design resulted in a biased estimate of the association between thromboembolic events and COVID-19. Patients visiting the hospital, almost always received a COVID-19 (PCR or antigen) test, especially early in the pandemic, while patients who did not visit the hospital were subject to underreporting infection data. This underreporting (or misclassification of infected as uninfected) led to an inflated IRR of thromboembolic events after COVID-19.
We proposed a simple and efficient method to quantify the association between thromboembolic events and COVID-19 while dealing with the misclassification issue. The main idea is to select a subset of control (i.e., subjects without thromboembolic events) who had a hospital visit for reasons independent of COVID-19 and therefore had complete infection data. To this end, we used patients who had a diagnosis code for physical injury (see Supplementary Table 2 ) at a hospital visit as the control group, since we would not expect any causal association between physical injury and COVID-19. We used a case-control design, in which patients with a thromboembolic event are considered as cases, and patients with a physical injury are considered as controls. If an individual had multiple hospital visits for thromboembolic events or physical injuries, we considered only the first visit. As physical injuries can be risk factors for thromboembolic events 26 , 27 , we therefore excluded patients who experienced both events at the same visit. We determined the COVID-19 status based on the COVID-19 test results during the 28 days prior to the date of the event (Supplementary Fig. 2 ). If an individual had a positive test result, this subject was classified as exposed to COVID-19, otherwise, unexposed. We compared the odds of infection (exposed) vs no infection (unexposed) in the cases (with thromboembolic events) vs controls (with physical injury) using a logistic regression model adjusted for age, race, gender, Charlson comorbidity index (CCI), number of visits, and prior vaccination status (yes/no). Patients who had any COVID-19 vaccine between the date of the positive COVID-19 test and the date of the event were removed. The number of visits was fit with a natural spline with three degrees of freedom. The CCI was obtained using the R package comorbidity and categorized into four categories, ‘0’, ‘1–2’, ‘3–4’, and ‘ ≥ 5’ 28 , 29 . Analyses were done after excluding patients with incomplete covariate data.
COVID-19 vaccines are protective against COVID-19 and COVID-19 severity 30 , 31 , 32 , and so can indirectly decrease the likelihood of experiencing a thromboembolic event. Hence, we conducted a risk-benefit analysis to estimate the net RR of thromboembolic events after vaccination by considering the role of vaccination in preventing infection-associated thromboembolic events. Figure 1 illustrates the direct and indirect effect of the COVID-19 vaccination on the occurrence of thromboembolic events while considering vaccine efficacy (VE). As presented in the diagram, the association between thromboembolic events and COVID-19 vaccination is described by two paths, the direct association between thromboembolic events and vaccination, and the indirect association between thromboembolic events and vaccination via potential reduction in the risk of thromboembolic events through decreasing the risk of COVID-19. We estimated the overall influence of vaccination on the occurrence of thromboembolic events by considering both direct and indirect paths.
COVID-19 (I), individuals with COVID-19. COVID-19 vaccination (V), individuals with COVID-19 vaccines. Thromboembolic events (Y), individuals with thromboembolic events. V → I indicates vaccine effect (VE) in preventing COVID-19, V → Y indicates the risk of thromboembolic events after COVID-19 vaccination, I → Y indicates the risk of thromboembolic events after COVID-19, V → Y (via I) indicates the risk of thromboembolic events after vaccination accounting for vaccine effect in reducing infection-associated thromboembolic events.
Let \({\rm{P}}\left({\rm{I}}|{\rm{V}}\right)\) and \({\rm{P}}\left({\rm{I}}|\bar{{\rm{V}}}\right)\) be the probability of COVID-19 ( \({\rm{I}})\) in vaccinated ( \({\rm{V}}\) ) and unvaccinated ( \(\bar{{\rm{V}}}\) ) subjects, respectively. Let \({\rm{P}}\left({\rm{Y}}|\bar{{\rm{V}}},\bar{{\rm{I}}}\right),{\rm{P}}\left({\rm{Y}}|{\rm{V}},\bar{{\rm{I}}}\right),{\rm{P}}\left({\rm{Y}}|{\rm{I}},\bar{{\rm{V}}}\right),\) and \({\rm{P}}\left({\rm{Y}}|{\rm{I}},{\rm{V}}\right)\) be the probability (or risk) of thromboembolic events ( \({\rm{Y}})\) in unvaccinated and uninfected, vaccinated and uninfected, unvaccinated and infected, and vaccinated and infected subjects, respectively.
With the above notations, for a vaccinated subject, the total risk of thromboembolic events is \({\rm{P}}\left({\rm{Y}}|{\rm{V}},\bar{{\rm{I}}}\right)+{\rm{P}}\left({\rm{I}}|{\rm{V}}\right)\times {\rm{P}}\left({\rm{Y}}|{\rm{I}},{\rm{V}}\right)\) , where the product \({\rm{P}}\left({\rm{I}}|{\rm{V}}\right)\times {\rm{P}}\left({\rm{Y}}|{\rm{I}},{\rm{V}}\right)\) is the indirect risk calculated by multiplying the risk of COVID-19 of a vaccinated subject and the risk of thromboembolic events given a COVID-19 in the vaccinated group. Similarly, the overall risk of thromboembolic events for an unvaccinated subject is given by \({\rm{P}}\left({\rm{Y}}|\bar{{\rm{V}}},\bar{{\rm{I}}}\right)+{\rm{P}}\left({\rm{I}}|\bar{{\rm{V}}}\right)\times {\rm{P}}\left({\rm{Y}}|{\rm{I}},\bar{{\rm{V}}}\right)\) . Hence the net RR ( \({{\rm{RR}}}_{{\rm{Net}}}\) ) of thromboembolic events for a vaccinated subject compared to an unvaccinated subject is
The terms \({{\rm{RR}}}_{{\rm{V}}}\) is the RR of thromboembolic events comparing vaccinated versus unvaccinated in subjects without COVID-19, and \({{\rm{RR}}}_{{\rm{I|}}\bar{{\rm{V}}}}\) is the RR of thromboembolic events comparing subjects with and without COVID-19 in the unvaccinated group. The term \({{\rm{RR}}}_{{\rm{IV}}}\) is the RR of thromboembolic events in subjects who have both vaccination and infection, compared to the group of subjects who do not have any exposures.
We further defined VE as \({\rm{VE}}=1-{\rm{P}}({\rm{I|V}})/{\rm{P}}({\rm{I|}}\bar{{\rm{V}}})\) , then plugged VE into Eq. (1) to obtain
If \({{\rm{RR}}}_{{\rm{Net}}}\) is smaller than one, COVID-19 vaccination offers protection against thromboembolic events, with a lower \({{\rm{RR}}}_{{\rm{Net}}}\) implying a stronger protection.
Statistical analyses were performed in R 4.3.0. We reported odds ratio (OR) and IRR with 95% CIs and p -values from the two-sided test. We generated a figure for \({{\rm{RR}}}_{{\rm{Net}}}\) over a range of VE values based on the estimates of ORs and IRRs.
We used de-identified EHR data, the use of which was approved by the Institutional Review Board of Corewell Health.
During the study period from December 1st, 2020, to August 31st, 2022, there were 747,070 subjects at Corewell Health who received mRNA-based vaccines, among which 279,229 (37.38%) had the primary series of mRNA-1273 and 467,841 (62.62%) took BNT162b2. Overall, the number of fully vaccinated patients was 711,460 (95.23%), and 35,610 (4.77%) patients received only one dose. The median age was 57 (with interquartile range [IQR]: 40–69), and 59.81% of patients were female. There were 367,105 patients taking at least one COVID-19 test (antigen or PCR), among which 78,568 (21.4%) patients received positive results. The median age was 52 (with interquartile range [IQR]: 34–67), and 61.44% of patients were female.
In the study cohort of vaccination exposure, there were 16,640 patients who had at least one thromboembolic event and had the first dose of either mRNA-1273 or BNT162b2 vaccine. Patient demographics are presented in Table 1 . We identified 2724 events in the control period, 722 events within 28 days after the first dose, and 786 events within 28 days after the second dose.
In the study cohort of COVID-19 exposure, there were 18,004 patients who had a thromboembolic event (cases) and 88,139 patients who had a physical injury (controls) at a hospital visit. 16.96% of cases and 1.48% of controls had COVID-19 within 28 days before the event. Demographics of patients are presented in Table 2 .
Based on the SCCS analysis, we found an increased risk of thromboembolic events 28 days after the first dose (IRR = 1.19, 95% confidence interval (CI): [1.08, 1.31], p -value < 0.001), and after the second dose (IRR = 1.22, 95% CI: [1.11, 1.34], p -value < 0.001) of the mRNA-based vaccines.
We studied the risk of thromboembolic events in a 28-day window after vaccination based on prior research 8 . An event that occurs in a short period (such as 28 days) is more likely to be attributable to the vaccines. We also conducted a sensitivity analysis using a 60-day window after vaccination. The conclusions remained the same with slightly lower IRRs (IRR = 1.13, 95% CI: [1.03, 1.24] after the first dose, and IRR = 1.14, 95% CI: [1.05, 1.3] after the second dose).
Supplementary Figs. 3 and 4 show the IRRs for subgroup analyses by age (“18–31”, “31–50”, and “≥51”) and gender (female/male). We found that the effects of vaccination on thromboembolic events were similar between age groups and gender groups.
Naïve SCCS analysis showed a very large increased risk of thromboembolic events associated with COVID-19 (IRR = 19.36, 95% CI: [17.64, 21.26], p -value < 0.001). However, a similar analysis using the physical injury as an event also derived a large increased risk (IRR = 3.31, 95% CI: [3.10, 3.54], p -value < 0.001), indicating misclassification bias as COVID-19 should not substantially increase the risk of physical injury. In the case-control analysis with controls having a physical injury, we found that COVID-19 increased the risk of thromboembolic events but with a much smaller magnitude than the risk in the SCCS analysis (although it is still larger than the vaccination exposure). Moreover, the degree of the increased risks was modified by vaccination status (Fig. 2 ). The reported OR for the unvaccinated group was 4.65 (95% CI: [4.18, 5.17], p -value < 0.001) compared to 2.77 (95% CI: [2.40, 3.24], p -value < 0.001) for the vaccinated group. We observed the increased risks of thromboembolic events after COVID-19 in both groups, but vaccination appears to confer some protection against infection-associated thromboembolic events, given the lower OR. Alternatively, we divided the vaccinated group into four categories based on the time to the last vaccination (“≥365 days”, “180–365 days”, “90–180 days”, and “<90 days”). The effects of COVID-19 on thromboembolic events were similar across the four vaccinated groups. The results are in Supplementary Fig. 5 .
OR is denoted by a solid circle and a 95% CI is represented by a line. The x -axis is plotted on the natural log scale. CCI Charlson comorbidity index. Infection or non-infection refers to COVID-19.
We also conducted two sensitivity analyses. In the first analysis, rather than adjusting for the CCI, we adjusted individual risk factors that might be related to a thromboembolic event. These are congestive heart failure, peripheral vascular disease, cerebrovascular disease, chronic pulmonary disease, diabetes with complications, cancer, moderate or severe liver disease, and metastatic solid tumors. We included the above eight risk factors (present or absent) in the logistic regression model. The effect of COVID-19 on the outcome of thromboembolic events was similar to the analysis with CCI. Results can be found in Supplementary Fig. 6 .
We assumed that patients who visited hospitals were routinely tested for COVID-19, especially during the early pandemic. Based on Corewell Health’s policy, patients who visited the healthcare system before March 1st, 2022, were tested for COVID-19. In our study cohort, 74.05% of participants had a hospital visit before March 1st, 2022. We conducted a sensitivity analysis using only these patients and the conclusions remained the same. See results in Supplementary Fig. 7 .
Our analysis in the previous sections gave an IRR of 1.22 as the measure of the association between thromboembolic events and the second dose of COVID-19 vaccination, therefore, we set \({{\rm{RR}}}_{{\rm{V}}}\) = 1.22. We also obtained odd ratios \({{\rm{OR}}}_{{\rm{I|}}\bar{{\rm{V}}}}\) = 4.65 and \({{\rm{OR}}}_{{\rm{IV}}}\) = 2.82 from the analysis using the case-control design. Since the RR is very close to the OR when the event is rare, we therefore set \({{\rm{RR}}}_{{\rm{I|}}\bar{{\rm{V}}}}\) = 4.65 and \({{\rm{RR}}}_{{\rm{IV}}}\) = 2.82, as the thromboembolic events are rare 33 . Hence, plugging these estimators into Eq. (2), the \({{\rm{RR}}}_{{\rm{Net}}}\) becomes
Figure 3 illustrates the \({{\rm{RR}}}_{{\rm{Net}}}\) of thromboembolic events after COVID-19 vaccination as a function of VE. As VE increases from 0 to 1, \({{\rm{RR}}}_{{\rm{Net}}}\) decreases and reaches a point where vaccine benefits outweigh the harms. Specifically, vaccines with higher VE offer higher protection against thromboembolic events. For example, the effectiveness of mRNA-based COVID-19 vaccines against infection was 61% during the Delta period and 46% during the Omicron period 34 , 35 , 36 . Given an infection rate of 0.08 among unvaccinated subjects, the risk of thromboembolic events was decreased by 4.62% in the Delta period, which is higher than 2.07% in the Omicron period. Moreover, vaccines offer stronger protection during periods with higher infection rates. For example, with the infection rate of 0.1 in unvaccinated subjects, the reduction of the risk of thromboembolic events was higher (by 9.19% in Delta and 6.23% in the Omicron period), compared to the scenario when the infection rate was 0.08.
The x -axis is VE, and the y -axis is the net RR of thromboembolic events.
The list of ICD-10 codes for thromboembolic events is based on a previous publication 8 , including old myocardial infarction (I252). Old myocardial infarction (I252) reports for any myocardial infarction described as older than four weeks. However, our study cohort removed subjects with an I252 code who had any thromboembolic event with ICD-10 codes listed in Table S1 in the prior year. Therefore, we can consider observing I252 in the study period as a new incidence. There were 20,002 (18.84%) patients with a hospital visit associated with the I252 code. We conducted a sensitivity analysis by excluding these patients and the conclusions did not change. The estimated IRRs of thromboembolic events are 1.16 and 1.17 after vaccine dose 1 and dose 2, respectively, which are slightly smaller than the original results including the I252 code (IRRs were 1.19 and 1.22 after the first and second dose). The association between COVID-19 and thromboembolic events is higher in the unvaccinated group (OR = 5.77 without I252 and OR = 4.65 with I252) and similar in the vaccinated group (OR = 2.80 without I252 and OR = 2.77 with I252). Hence, given the same infection rate and VE, vaccination offered a stronger protection, compared to the analysis with the I252 codes. For example, given an infection rate in the unvaccinated population of 0.08 and a VE of 0.8, vaccination lowers the risk of thromboembolic events by 17.14% without I252, compared to 6.67% in the analysis with I252. Detailed results are in Supplementary Figs. 8 and 9 . We considered the analysis that includes the I252 code as the main analysis to represent more conservative results.
We found that both COVID-19 vaccination and COVID-19 increase the risk of thromboembolic events. However, evidence implies that the likelihood of experiencing a thromboembolic event after COVID-19 is much higher than after vaccination. Our analysis agrees with previous research, indicating that COVID-19 is a more dangerous risk factor for thromboembolic events than vaccination 8 , 11 , 12 , 13 , 14 .
Different from existing work, we evaluated the association between thromboembolic events and COVID-19 using a case-control study, avoiding the misclassification issue associated with the SCCS design. We also studied the effect of prior vaccination on reducing infection-associated thromboembolic events. Moreover, we included both COVID-19 vaccination and COVID-19 in the analysis of the risk of thromboembolic events and conducted a risk-benefit analysis by comparing the magnitude of the increased risk through the direct effect of COVID-19 vaccination with the reduced risk through the indirect pathway via protection against severe diseases. Our analysis provides evidence that COVID-19 vaccination directly increases the risk of thromboembolic events, but indirectly reduces the risk of infection-associated events. Results show that the indirect benefit of preventing infection-associated thromboembolic events outweighs the direct harm if the VE and infection rate reaches certain levels. Moreover, COVID-19 vaccination may have additional benefits in preventing thromboembolic events associated with COVID-19, as a higher rate of vaccination increases the overall level of immunity in the population, reducing the spread of the virus and conferring collective protection against infection-associated thromboembolic events and other health risks associated with COVID-19.
There are several limitations to this study. First, using ICD-10 codes to identify thromboembolic events may be subject to phenotype errors. Second, Corewell Health has 22 hospitals, and the catchment area for these hospitals is across many counties, hence patients may seek care at other facilities outside the Corewell Health system, leading to missing data such as infection data. To deal with the missing infection data, we used the case-control study. Moreover, the use of a prior number of hospital visits as covariates in the regression model mitigates the bias due to differing degrees of interaction with the Corewell Health system between infected and control subjects. However, patients with a hospital visit due to injuries may not be the perfect control group, but it is clearly better than a control group of patients without thromboembolic events. Therefore, we may not totally correct the bias, but we reduce it. Finally, the study population for vaccine doses 1 and 2 are different. If a subject had a thromboembolic event after the first vaccine dose, this subject is unlikely to receive the second dose, therefore, the population who received the second dose only includes subjects who did not have a thromboembolic event after the first dose.
Despite these limitations, our study makes a critical contribution to quantifying the net risk of thromboembolic events associated with COVID-19 vaccination. It accounts for both the direct effects of vaccination and the indirect effects of protection against COVID-19 and severe diseases. The dual consideration is vital for a comprehensive understanding of the risk-benefit profile. The mechanism of vaccination is to simulate the immune response the body has against infection using a dead/attenuated virus or mRNA, which can lead to side effects similar to those of the virus, albeit in a less severe form (e.g., thromboembolic events, myocarditis 37 , acute kidney injury 38 , 39 ). Our finding highlights the necessity of evaluating both the indirect benefits and direct harms of vaccination to provide a complete and accurate assessment of vaccine safety. This comprehensive approach ensures a balanced understanding of the risks and the benefits, reinforcing the overall safety and efficacy of vaccination programs.
Our risk-benefit analysis was conducted on the population level. This analysis can also be stratified by patient groups of interest. For example, the risk-benefit of vaccination might be different between older and younger populations. Moreover, our findings are for a broad range of thromboembolic conditions, so more research is needed on the specific biological mechanisms connecting COVID-19 and mRNA vaccination to these events, both to establish causality and help identify a more specific set of conditions or risk factors.
The datasets analyzed during the current study are not publicly available due to privacy or ethical restrictions.
Code for this study is available from the corresponding author on request.
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We thank Kevin Heinrich at Quire and Martin Witteveen-Lane for querying the data from the Corewell Health Epic system. This study was funded by the National Institute of Allergy and Infectious Diseases of the National Institutes of Health under award number R01AI158543. The funder played no role in the study design, data collection, analysis, and interpretation of data, or the writing of this manuscript.
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Division of Biostatistics & Health Informatics, Corewell Health Research Institute, Royal Oak, MI, USA
Huong N. Q. Tran
Department of Biostatistics, University of Michigan, Ann Arbor, MI, USA
Malcolm Risk
William Beaumont University Hospital, Corewell Health East, Royal Oak, MI, USA
Girish B. Nair
Division of Biostatistics, Department of Preventive Medicine, Northwestern University, Chicago, IL, USA
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H.N.Q.T.: manuscript writing, study design, statistical analysis, and data preparation. M.R.: manuscript writing and study design. G.B.: clinical advice and study design. L.Z.: manuscript writing, method development, study design, and statistical analysis.
Correspondence to Lili Zhao .
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Tran, H.N.Q., Risk, M., Nair, G.B. et al. Risk benefit analysis to evaluate risk of thromboembolic events after mRNA COVID-19 vaccination and COVID-19. npj Vaccines 9 , 166 (2024). https://doi.org/10.1038/s41541-024-00960-7
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Received : 06 May 2024
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Published : 13 September 2024
DOI : https://doi.org/10.1038/s41541-024-00960-7
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2. Phillip Cummings Stole 33,000 Credit Reports Which Cost Victims Between $50-100 Million. At the time it happened, Philip Cumming was responsible for the largest identity theft case in US history. Cummings worked as a help desk for a software company in Long Island.
Case Study #2: Dr. Jubal Yennie. As demonstrated by the above incident, it doesn't take much information to impersonate someone via social media. In the case of Dr. Jubal Yennie, all it took was a name and a photo. In 2013, 18-year-old Ira Trey Quesenberry III, a student of the Sullivan County School District in Sullivan County, Tennessee ...
Seattle - A former Seattle resident who defrauded federal COVID-19 benefit programs of more than $1 million pleaded guilty today in U.S. District Court in Seattle, announced U.S. Attorney Nick Brown. Bryan Alan Sparks, 42, was indicted for the fraud scheme in November 2021. Today Sparks pleaded guilty to wire fraud and aggravated identity theft.
9 Unbelievable Identity Theft Stories. 1. The "Tinder Swindler" who scammed lonely lovers out of millions. Shimon Hayut is the subject of the Netflix documentary The Tinder Swindler, and he's one of the most brazen scammers on this list. Simon Hayut (aka the Tinder Swindler). Source: The Sun.
11 Nightmare Identity Theft Cases. We asked our community for identity theft case examples. Unfortunately, we got plenty of horrifying stories in return. Below are some recent identity theft stories from real people. "My first experience with stolen identity happened when my parents told me to get a credit card.
Identity theft is a pervasive issue that affects millions of Americans each year, with the Federal Trade Commission reporting approximately 41.4 million victims in 2022 alone. The average financial loss per victim in the same year was about $7,697, highlighting the significant impact this crime can have on individuals' financial well-being. Moreover, the rise in …
For 76% of identity-theft victims in 2021, the most recent incident involved the misuse of only one type of existing account, such as a credit card or bank account. About 59% of identity-theft victims had financial losses of $1 or more that totaled $16.4 billion in 2021. In 2021, about 2% of persons age 16 or older experienced the misuse of an ...
Financial loss for all identity theft. Across all incidents of identity theft reported in 2021, about 59% of victims experienced a financial loss of $1 or more (table 5). These victims had financial losses totaling $16.4 billion. The mean loss was $1,160 per victim, and the median loss was $200. TABLE 5.
A cloned account mimicking that of freelance model Elle Jones offered pornographic content through a link. One of the problems with social media identity theft is that the reporting mechanisms are ...
reports of identity theft has increased over the study time period. From 2000 to 2001, identity theft jumped from 112 to 230 - a 105% increase. Over the same time period, credit card fraud increased 43%, motor vehicle theft increased 13%, robbery remained. stable, and check fraud decreased 32%.
OVERVIEW. The results of Javelin's 2020 Identity Fraud Survey serve as a wake-up call—one that will force financial institutions, businesses, and the payment industry to reevaluate how identity fraud is managed. Total identity fraud reached $16.9 billion (USD) in 2019, yet the dollar loss is only part of the story.
In my case, my credit card company thwarted my identity thieves before they made a single charge. I did, however, lose several days of my life trying to shore up my compromised identity.
In these consolidated appeals involving Appellants' convictions for identity theft, the Court of Appeals held (1) the law defines the use of personal identifying information of another as one of the express means by which a defendant assumes that person's identity; and (2) therefore, the People may establish that a defendant "assumes the identity of another" within the meaning of New ...
Datos Insights. Boston, March 9, 2021 - Identity theft is a growing problem in the U.S. In 2019, losses from identity theft cases were US$502.5 billion and rapidly increased to US$712.4 billion in 2020, a 42% increase year-over-year. Identity theft losses grew very rapidly in 2020 (and will continue in 2021) due to the very high rate of ...
Police reports are critical for victims to pursue an identity theft case (OVC, 2010). For victims of certain forms of identity theft, the discovery of victimization can take as long as 6 months or more (Synovate, 2003, 2007). In cases where personal information is exposed due to data breaches, victims might have greatly varying experiences of ...
I focus on the case of identity theft, showing how that event—experienced by tens of millions of Americans annually—contributes to economic insecurity. ... Huff Rodney, Kane John. 2010. "Differentiating Identity Theft: An Exploratory Study of Victims Using a National Victimization Survey." Journal of Criminal Justice 38(5):1045-52 ...
A case study methodology was selected for this project. The results indicate that the identity theft trend is different than the trends for other theft related offenses -- credit card fraud, check fraud, robbery and motor vehicle theft. The data suggest that identity theft is increasing more rapidly than the other theft orientated offenses.
Abstract. Identity theft victimization is associated with serious physical and mental health morbidities. The problem is expanding as society becomes increasingly reliant on technology to store and transfer personally identifying information. Guided by lifestyle-routine activity theory, this study sought to identify risk and protective factors ...
The process of restoring one's financial standing after identity theft is time-consuming and stressful. 5. Essential Measures to Protect Against Identity Theft in the Digital Era 5.1 Safeguarding Personal Information: Best Practices for Individuals. In the digital era, protecting personal information is crucial to prevent identity theft.
consequences of identity theft. A th reat t hat has emerged with the development of Internet. services is undoubtedly the ease of obtaining s hort-term. financial support - via avai lable Internet ...
The students demonstrated a significant misunderstanding of who perpetrators typically were targeting when stealing personal information or what perpetrators of identity theft were looking for. , - The results of the study contribute to a better understanding of the students' knowledge about the risks associated with identity crime.
However, because the case was brought under the FCRA, it is unclear whether the holding applies to all identity theft cases or just to those prosecuted under FCRA. 4. In fact, many police departments in the United States, until recently, were not equipped to investigate identity theft and would even refuse to take a report unless the actual ...
Flores is charged with 10 counts of health care fraud, each carrying a possible 10-year-maximum sentence and up to a $250,000 fine. He is also facing three counts of aggravated identity theft. If convicted, he faces a mandatory two years in federal prison which must be served consecutively to any other sentence imposed.
A Mt. Pleasant man who pleaded guilty to identity theft and bank fraud has been sentenced to serve a year and a day in prison. United States District Judge Thomas L. Ludington on Thursday also ...
Former Rep. George Santos was expelled from Congress after questions were raised about the New York Republican's resume and his use of campaign funds.
The net vaccine effect was estimated using results from SCCS and case-control studies. We used electronic health record data from Corewell Health (16,640 subjects in SCCS and 106,143 in case-control).