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Original Contribution |

Civil Protection Orders and Risk of Subsequent Police-Reported Violence FREE

Victoria L. Holt, PhD, MPH; Mary A. Kernic, PhD, MPH; Thomas Lumley, PhD; Marsha E. Wolf, PhD; Frederick P. Rivara, MD, MPH
[+] Author Affiliations

Author Affiliations: Department of Epidemiology (Drs Holt, Kernic, Wolf, and Rivara) and Department of Biostatistics (Dr Lumley), School of Public Health and Community Medicine, University of Washington; Department of Pediatrics, School of Medicine, University of Washington (Dr Rivara); and Harborview Injury Prevention and Research Center, Seattle, Wash (Drs Holt, Kernic, Wolf, and Rivara).


JAMA. 2002;288(5):589-594. doi:10.1001/jama.288.5.589.
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Context Approximately 1.5 million US women experience intimate partner violence annually. Approximately 20% of these women obtain civil protection orders, but the effectiveness of such orders in preventing future violence is unclear.

Objective To assess associations between obtaining a protection order and risk of subsequent police-reported intimate partner violence.

Design, Setting, and Subjects Retrospective cohort study of 2691 adult female residents of Seattle, Wash, with an incident of male intimate partner violence reported to the Seattle Police Department between August 1, 1998, and December 31, 1999.

Main Outcome Measure Relative risk (RR) of police-reported physical and psychological abuse in the 12 months following the index incident according to protection order status (temporary protection order, usually in effect for 2 weeks; permanent protection order, usually in effect for 12 months; or no protection order).

Results Overall rates of police-reported physical and psychological abuse in the 12 months of follow-up were 13.5 per 100 person-years and 12.3 per 100 person-years, respectively. After controlling for cohabitation at time of index incident and index incident offense type, women with temporary protection orders in effect were more likely than women without protection orders to be psychologically abused (RR in the first 6 months after the index incident, 4.0; 95% confidence interval [CI], 2.2-7.2; RR in the entire 12 months after the index incident, 4.9; 95% CI, 2.8-8.6), while women with permanent protection orders in effect were less likely than those without orders to be physically abused (RR in the first 6 months, 0.4; 95% CI, 0.1-1.1; RR in the entire 12 months, 0.2; 95% CI, 0.1-0.8).

Conclusions Permanent, but not temporary, protection orders are associated with a significant decrease in risk of police-reported violence against women by their male intimate partners.

Intimate partner violence (IPV) is a frequent occurrence in the United States, with nearly 5 million physical or sexual assaults by intimate partners experienced by approximately 1.5 million women annually.1 In addition to injury-related visits, abused women have high frequencies of emergency department visits and hospitalizations for somatic and psychiatric diagnoses related to stress, including functional gastrointestinal disorders, loss of appetite, chest pain, headaches, anxiety, insomnia, alcohol abuse or dependence, post-traumatic stress disorder, depression, and suicide attempts.25

Several strategies can be used by abused women in an attempt to deter future violence, but limited financial and community resources such as battered women's shelters may restrict women's options. One widely available option is to obtain a civil protection order, a legally binding court order that restrains an individual who has committed an act of violence against a person from further acts against that person.6 Specifically, a protection order can prohibit the abuser from committing acts of violence; exclude the abuser from the residence shared by the petitioner and abuser; prohibit the abuser from harassing or contacting the petitioner by mail, telephone, or in person; award temporary custody of minor children; establish temporary visitation and restrain the abuser from interfering with custody; prohibit the abuser from removing the children from the jurisdiction of the court; and order the abuser to participate in treatment or counseling. Although approximately 20% of US women experiencing IPV obtain civil protection orders, their effectiveness in preventing IPV recurrence is unclear, and it has been suggested that they may in fact aggravate violence under certain conditions.710 While many case series have described the experiences of women with protection orders, only 1 published study has investigated protection order effectiveness by comparing abused women with and without protection orders. This interview-based study reported that violence frequency was not significantly affected by the presence of a civil restraining order, but the study's small size, low response rate, and short follow-up period limit this interpretation.11 The current study addresses this issue using linked data from a large population of women in an entire US city on whom criminal justice system information was available.

In this retrospective cohort study, subjects were all 2691 female residents of Seattle, Wash, who had a police-reported episode of IPV inflicted by a male former or current intimate partner between August 1, 1998, and December 31, 1999, and who had not obtained a permanent protection order in the prior 12 months. We obtained names of abused women from the Seattle Police Department Domestic Violence Unit database of all IPV incident reports and ascertained protection order status using information from the King County, Washington District and Superior Court records of filings for civil protection order. In King County, women seeking protection orders petition (free of charge) first for a temporary protection order, which is granted by a judge or commissioner for a period of 2 weeks. During these 2 weeks the abuser is served with both the petition and the temporary order, with notice of the date set for a hearing (approximately 2 weeks after the initial petition), at which time the court grants or denies a "permanent" protection order effective for 1 year or more. Approximately 57% of women in King County who file temporary protection orders against male intimate partners go on to obtain permanent orders. Anytime prior to the expiration date of a permanent protection order, the petitioner may return to court to request that the order be modified or terminated. Our study protocols were approved by the University of Washington Human Subjects Review Committee and the Washington State Department of Health Human Research Review Board.

The primary outcome in these analyses, subsequent police-reported abuse of a study subject by the same abuser, was ascertained from police-reported incidents of IPV during the 12 months following the initial police-reported incident (the index incident). Using police reports, we categorized subsequent IPV incidents as those including physical abuse (assault, reckless endangerment, or unlawful imprisonment) and those including psychological abuse (harassment, menacing, stalking, threats, disturbance, criminal trespass, custodial interference, interfering with IPV reporting, or property damage). Incidents in which the sole offense was a protection order violation were not included in our analyses. Additionally, we used Washington State Vital Statistics data to ascertain deaths during follow-up.

Demographic differences between women who obtained a temporary protection order at any time in the 12 months following the index incident (without a subsequent permanent order), women who obtained a permanent protection order at some time during the 12 months of follow-up, and women who obtained neither type of order at any time during the follow-up were assessed using χ2 tests, with P<.05 denoting significance.

The relative risk (RR) of subsequent police-reported IPV according to protection order status was estimated using Cox proportional hazards regression.12 Time to abuse (defined as a police-reported IPV incident during follow-up) was modeled as a function of time from entry into the cohort (the date of the index IPV incident). In all models, protection order status was modeled as a time-dependent variable, allowing subjects to change exposure categories as protection orders were initiated or terminated. Temporary protection orders were usually granted for a 2-week period, and permanent protection orders were for a 12-month period. The time from the filing of a temporary protection order until the order's typical automatic expiration 2 weeks later (or until the temporary protection order was rescinded if that came first) was counted as temporary protection order–exposed time. The time from the filing of a permanent protection order until the end of that woman's follow-up period (or until the permanent protection order was rescinded if that came first) was counted as permanent protection order–exposed time. Time during which a woman had neither a temporary protection order nor a permanent protection order in effect was counted as unexposed time. Two comparisons were made: (1) temporary protection order compared with no protection order, and (2) permanent protection order compared with no protection order. We allowed multiple incidents per subject, adjusting the SEs for dependencies between incident times, using Stata statistical software for all analyses.13 We calculated 2 sets of risk estimates of subsequent IPV according to protection order status: (1) for the first 6 months of follow-up after the index incident, and (2) for the entire 12 months of follow-up.

In multivariate models of the effect of protection orders on subsequent physical and psychological IPV, we considered as potential confounders the following baseline covariates that we previously found to be related to obtaining a protection order: subject age, pregnancy status, and alcohol and other drug use; abuser age and alcohol and other drug use; subject/abuser relationship, cohabitation at time of index incident; number of police-reported IPV incidents in the previous 12 months; and the type of offense at the index IPV incident (threat, weapon threat, physical assault, assault with weapon, sexual assault, injury).14 Covariates were entered into the regression models if they changed the risk estimates by 10% or more15; only cohabitation at time of index incident and index incident offense type met this criterion. We tested for the interaction of physical abuse during a temporary protection order and the effect of a permanent protection order using the likelihood ratio test; no significant (P <.05) interaction was noted.

Study subjects who did not obtain any protection orders, those who obtained only temporary protection orders, and those who obtained permanent protection orders during the 12 months following the index incident were similar in terms of age, pregnancy status, and IPV history with their abusers; but subjects who did not obtain protection orders during the follow-up were significantly more likely than other subjects to have used alcohol or other drugs at the index incident, as were their abusers (Table 1). Additionally, subjects who did not obtain protection orders were less likely than other women to have ever been married to their abusers and more likely to be living with them at the time of the index incident. The use of a weapon in the index incident did not differ by protection order status, but subjects who did not obtain protection orders were more likely than other women to have been assaulted or injured during the index incident.

Table Graphic Jump LocationTable 1. Subject, Abuser, and Relationship Characteristics at Time of Index Incident, by Protection Order Status

In the first 6 months of follow-up there were 222 incidents of police-reported physical abuse (16.5 incidents per 100 person-years) and 223 incidents of police-reported psychological abuse (16.6 incidents per 100 person-years). Over the entire 12 months of follow-up, there were 363 incidents of police-reported physical abuse (13.5 per 100 person-years) and 330 incidents of police-reported psychological abuse (12.3 per 100 person-years).

In the first 6 months of follow-up, the rate of police-reported physical abuse during times in which no protection order was in effect was 17.2 per 100 person-years, and the corresponding rates for temporary protection order–exposed time and permanent protection order–exposed time were 14.7 per 100 person-years and 5.4 per 100 person-years, respectively (Table 2). In time-dependent Cox proportional hazards regression models controlling for cohabitation and index incident offense type, the RR of police-reported physical abuse during the first 6 months after the index incident associated with a temporary protection order was 0.8 (95% confidence interval [CI], 0.2-3.4), and the risk associated with a permanent protection order was 0.4 (95% CI, 0.1-1.1) compared with no protection order. In the first 6 months of follow-up, the rate of police-reported psychological abuse during times in which no protection order was in effect was 16.0 per 100 person-years, and the corresponding rates for temporary protection order–exposed time and permanent protection order–exposed time were 95.6 per 100 person-years and 16.2 per 100 person-years, respectively. The RR of police-reported psychological abuse during the first 6 months after the index incident associated with a temporary protection order was 4.0 (95% CI, 2.2-7.2), and the risk associated with a permanent protection order was 1.1 (95% CI, 0.5-2.3), compared with no protection order.

Table Graphic Jump LocationTable 2. Rates and Relative Risks (RRs) of Subsequent Police-Reported Intimate Partner Violence (IPV) Associated With a Temporary or Permanent Civil Protection Order in the 6 Months Following the Index IPV Incident

Over the entire 12 months of follow-up, the rate of police-reported physical abuse during times in which no protection order was in effect was 14.0 per 100 person-years, and the corresponding rates for temporary protection order–exposed time and permanent protection order–exposed time were 24.7 per 100 person-years and 2.9 per 100 person-years, respectively (Table 3). In time-dependent Cox proportional hazards regression models controlling for cohabitation and index incident offense type, the RR of police-reported physical abuse during the 12 months after the index incident associated with a temporary protection order was 1.6 (95% CI, 0.6-4.4), and the risk associated with a permanent protection order was 0.2 (95% CI, 0.1-0.8) compared with no protection order. In the entire 12 months of follow-up, the rate of police-reported psychological abuse during times in which no protection order was in effect was 11.8 per 100 person-years, and the corresponding rates for temporary protection order–exposed time and permanent protection order–exposed time were 104.9 per 100 person-years and 10.2 per 100 person-years, respectively. The RR of police-reported psychological abuse during the entire 12 months after the index incident associated with a temporary protection order was 4.9 (95% CI, 2.8-8.6), and the risk associated with a permanent protection order was 0.9 (95% CI, 0.5-1.7) compared with no protection order.

Table Graphic Jump LocationTable 3. Rates and Relative Risks (RRs) of Subsequent Police-Reported Intimate Partner Violence (IPV) Associated With a Temporary or Permanent Civil Protection Order in the 12 Months Following the Index IPV Incident

To address the possibility that our results were unduly influenced by repeated recurrences among a small group of women, we conducted a series of sub-analyses limited to at most 1 failure per subject. While the risk for psychological abuse associated with temporary protection order exposure moderated somewhat in these analyses (RR for 12 months of follow-up, 4.4; 95% CI, 2.3-8.2), all other findings remained essentially the same.

There were 5 deaths from homicide in the study cohort, for a rate of 1.9 per 1000 person-years. Homicide mortality rates did not differ significantly by protection order status.

In this population-based cohort of all women with an incident of IPV reported to Seattle police, the overall rates of police-reported recurrence of physical and psychological abuse in the following 12 months were 14 per 100 person-years and 12 per 100 person-years, respectively. These numbers are substantially lower than those reported in studies using convenience samples of women obtaining protection orders, which have generally found that one third to one half of abused women self-report physical abuse in follow-up periods ranging from 4 months to 1 year, and approximately half report psychological abuse.1619 The frequency of IPV recurrence we found was more in accord with results from a study by Carlson et al,20 also based on police records, in which 23% of women with protection orders reported physical violence to the police in 2 years of follow-up.

Our findings also differ from those of the only other published study to compare outcomes for women with and without protection orders. In an interview-based study of abused women who participated in a family violence demonstration program, Grau et al11 reported that the likelihood of any abuse or violence in 4 months of observation did not differ significantly by protection order status. The study provided follow-up violence information on only 170 of the 270 participants, however, raising concerns about the adequacy of the study's power and its internal validity.

In the current study we used police reports to ascertain IPV recurrence to eliminate any bias that may result from the possibly atypical nature of the subset of abused women who will agree to participate in an interview-based study. However, in so doing we captured only that portion of IPV that was police-reported, which crime victim surveys have estimated to be approximately 50% of IPV incidents.21 Use of police-reported IPV to represent all abuse implicitly assumes that women with and without protection orders are equally likely to report violence to police if it occurs and that police are equally likely to respond. Because Seattle police respond to all calls, and because our data were based on the incident reports completed after all responses, we have no reason to believe that police response or recording depended on protection order status. If the completeness of reporting of IPV incidents to police varied by protection order status, our results may be biased. While we have no information to indicate that differential reporting of IPV existed, we can postulate that because they had taken a formal legal step to acknowledge IPV against them and request its cessation, it is possible that women with protection orders were more likely to report new abuse to police. Reporting differences may have been a factor particularly in incidents involving psychological abuse, for which temporary protection orders were associated with a quadrupled risk in our study. We found that the primary psychological abuse offense was more likely to be harassment in incidents involving women with current protection orders than it was in incidents involving women without current orders. If psychological abuse was relatively overreported by women during temporary protection order–exposed times, our results would be an overestimate of the adverse effect of temporary protection order on this type of abuse. On the other hand, a temporary protection order might have restrained the abuser from inflicting physical abuse, with a consequent increase in psychological abuse. Our finding of a quadrupling of psychological abuse risk during the time of a temporary protection order indicates that the time shortly after the index incident, when most temporary protection orders are issued, may be one of exceptional volatility between the subject and her abuser. However, that we did not find a parallel increase in risk of physical abuse with temporary protection order exposure provides some evidence that prior concerns of increased violence associated with protection order filing may be unfounded.

We had no contact with the subjects in this study; therefore, it is possible that, unknown to us, some subjects moved out of the Seattle area during the 12 months following their index incidents and we were unable to ascertain their IPV recurrence. We do not know how likely this out-migration was, but information we collected for another purpose may help us to estimate the potential magnitude. In our recent interview-based study of protection order effectiveness among Seattle women with police- or court-reported IPV, we found that we were able to retain in the study for 12 months 83% of participants with a protection order, and 74% of participants without a protection order.22 If the participants who were lost to follow-up in that study left the Seattle area, the difference in out-migration by protection order status would indicate that our results are an overestimate of the adverse effects and an underestimate of the beneficial effects of protection orders. However, given the relatively small numbers, it seems likely that any effect would be small. A related possibility is that subjects were not exposed to the potential for recurrence of violence because of their abusers' incarceration related to the index incident. Because only about 5% of reported IPV results in conviction and incarceration in King County, we think this is not likely to be an important factor.

Police data on IPV incidents provide a limited number of demographic or explanatory variables; therefore, the possibility of incomplete control for confounding exists. In our analyses we examined as potential confounders several variables that we previously found to be associated with obtaining a protection order14; only 2 of these were confounders of the associations between protection order status and IPV risk. The lack of confounding may have resulted from our analytic method, which used a time-dependent exposure variable, allowing subjects to contribute observation time in each category they experienced during follow-up. For instance, a woman who obtained a temporary order 2 weeks after the index incident and then a permanent order 2 weeks later contributed 0.5 person-months of unexposed time, 0.5 person-months of temporary protection order–exposed time, and 11 person-months of permanent protection order–exposed time; and one who obtained a temporary order 2 weeks after the index incident and then no permanent order contributed 0.5 person-months of temporary protection order–exposed time and 11.5 person-months of time without protection order exposure. Therefore, some potentially confounding variables, such as study subjects' personal characteristics, may also have been distributed across protection order and nonprotection order exposure categories.

In this study we found that having a permanent protection order in effect was associated with a statistically significant 80% reduction in police-reported physical violence in the 12 months after an IPV incident. We controlled in our analyses for all variables that we found to be associated with a woman's likelihood of obtaining a civil protection order as well as the likelihood of future violence, but we may not have captured important characteristics that reflect a woman's motivation and ability to initiate and complete the process of obtaining a protection order as well as her resolve not to be abused further. Further comparative studies of abused women with and without protection orders that ascertain the determinants of the decision whether to seek an order, other concurrent steps taken to prevent violence recurrence, and women's opinions of the reasons for violence cessation or recurrence may help explain how to enhance the protective impact of civil protection orders.

Tjaden P, Thoennes N. Full Report of the Prevalence, Incidence, and Consequences of Violence Against Women. Washington, DC: US Dept of Justice; 2000.
Kernic MA, Wolf ME, Holt VL. Rates and relative risk of hospital admission among women in violent intimate partner relationships, King County, Washington.  Am J Public Health.2000;90:1416-1420.
McCauley J, Kern DE, Kolodner K, Derogatis LR, Bass EB. Relation of low-severity violence to women's health.  J Gen Intern Med.1998;13:687-691.
Campbell JC, Lewandowski LA. Mental and physical health effects of intimate partner violence on women and children.  Psychiatr Clin North Am.1997;20:353-374.
Resnick HS, Acierno R, Kilpatrick DG. Health impact of interpersonal violence, 2: medical and mental health outcomes.  Behav Med.1997;23:65-78.
Finn P, Colson S. Civil Protection Orders: Legislation, Current Court Practice, and Enforcement. Washington, DC: US Dept of Justice; 1990.
Tjaden P, Thoennes N. Extent, Nature, and Consequences of Intimate Partner Violence. Washington, DC: National Institute of Justice; 2000.
Glick B, Johnson S, Pham C. 1998 Oregon Domestic Violence Needs Assessment: A Report to the Oregon Governor's Council on Domestic Violence. Portland: Oregon Health Division and Multnomah County Health Dept; 1999.
 Use of medical care, police assistance, and restraining orders by women reporting intimate partner violence—Massachusetts, 1996-1997.  MMWR Morb Mortal Wkly Rep.2000;49:485-488.
Crowell NA, Burgess AW. Understanding Violence Against Women. Washington, DC: National Academy Press; 1996.
Grau J, Fagan J, Wexler S. Restraining orders for battered women: issues of access and efficacy. In: Schweber C, Feinman C, eds. Criminal Justice Politics and Women: The Aftermath of Legally Mandated Change. New York, NY: Haworth Press Inc; 1985:13-28.
Cox DR. Regression models and life tables.  J R Stat Soc Serv B.1972;34:187-202.
StataCorp.  Stata Statistical Software: Release 6.0. College Station, Tex: Stata Corp; 1999.
Wolf ME, Holt VL, Kernic MA, Rivara FP. Who gets protection orders for intimate partner violence?  Am J Prev Med.2000;19:286-291.
Maldonado G, Greenland S. Simulation study of confounder selection strategies.  Am J Epidemiol.1993;138:923-936.
Chaudhuri M, Daly K. Do restraining orders help? battered women's experience with male violence and legal process. In: Buzawa ES, Buzawa CG, eds. Domestic Violence: The Changing Criminal Justice Response. Westport, Conn: Auburn House; 1992:227-252.
Klein AR. Re-abuse in a population of court-restrained male batterers after two years: development of a predictive model. In: Buzawa ES, Buzawa CG, eds. Do Arrests and Restraining Orders Work? Thousand Oaks, Calif: SAGE Publications; 1996:192-213.
Harrell A, Smith B, Newmark L. Court Processing and the Effects of Restraining Orders for Domestic Violence Victims. Washington, DC: The Urban Institute; 1993.
Horton AL, Simonidis KM, Simonidis LL. Legal remedies for spousal abuse: victim characteristics, expectations, and satisfaction.  J Fam Viol.1987;2:265-279.
Carlson MJ, Harris SD, Holden GW. Protective orders and domestic violence: risk factors for re-abuse.  J Fam Viol.1999;14:205-226.
Bachman R, Saltzman LE. Violence Against Women: Estimates From the Redesigned Survey. Washington, DC: US Dept of Justice; 1995.
Wolf ME, Kernic MA, Holt, VL, Rivara FP. Do protection orders work?  Am J Epidemiol.2000;151(suppl):S53.

Figures

Tables

Table Graphic Jump LocationTable 1. Subject, Abuser, and Relationship Characteristics at Time of Index Incident, by Protection Order Status
Table Graphic Jump LocationTable 2. Rates and Relative Risks (RRs) of Subsequent Police-Reported Intimate Partner Violence (IPV) Associated With a Temporary or Permanent Civil Protection Order in the 6 Months Following the Index IPV Incident
Table Graphic Jump LocationTable 3. Rates and Relative Risks (RRs) of Subsequent Police-Reported Intimate Partner Violence (IPV) Associated With a Temporary or Permanent Civil Protection Order in the 12 Months Following the Index IPV Incident

References

Tjaden P, Thoennes N. Full Report of the Prevalence, Incidence, and Consequences of Violence Against Women. Washington, DC: US Dept of Justice; 2000.
Kernic MA, Wolf ME, Holt VL. Rates and relative risk of hospital admission among women in violent intimate partner relationships, King County, Washington.  Am J Public Health.2000;90:1416-1420.
McCauley J, Kern DE, Kolodner K, Derogatis LR, Bass EB. Relation of low-severity violence to women's health.  J Gen Intern Med.1998;13:687-691.
Campbell JC, Lewandowski LA. Mental and physical health effects of intimate partner violence on women and children.  Psychiatr Clin North Am.1997;20:353-374.
Resnick HS, Acierno R, Kilpatrick DG. Health impact of interpersonal violence, 2: medical and mental health outcomes.  Behav Med.1997;23:65-78.
Finn P, Colson S. Civil Protection Orders: Legislation, Current Court Practice, and Enforcement. Washington, DC: US Dept of Justice; 1990.
Tjaden P, Thoennes N. Extent, Nature, and Consequences of Intimate Partner Violence. Washington, DC: National Institute of Justice; 2000.
Glick B, Johnson S, Pham C. 1998 Oregon Domestic Violence Needs Assessment: A Report to the Oregon Governor's Council on Domestic Violence. Portland: Oregon Health Division and Multnomah County Health Dept; 1999.
 Use of medical care, police assistance, and restraining orders by women reporting intimate partner violence—Massachusetts, 1996-1997.  MMWR Morb Mortal Wkly Rep.2000;49:485-488.
Crowell NA, Burgess AW. Understanding Violence Against Women. Washington, DC: National Academy Press; 1996.
Grau J, Fagan J, Wexler S. Restraining orders for battered women: issues of access and efficacy. In: Schweber C, Feinman C, eds. Criminal Justice Politics and Women: The Aftermath of Legally Mandated Change. New York, NY: Haworth Press Inc; 1985:13-28.
Cox DR. Regression models and life tables.  J R Stat Soc Serv B.1972;34:187-202.
StataCorp.  Stata Statistical Software: Release 6.0. College Station, Tex: Stata Corp; 1999.
Wolf ME, Holt VL, Kernic MA, Rivara FP. Who gets protection orders for intimate partner violence?  Am J Prev Med.2000;19:286-291.
Maldonado G, Greenland S. Simulation study of confounder selection strategies.  Am J Epidemiol.1993;138:923-936.
Chaudhuri M, Daly K. Do restraining orders help? battered women's experience with male violence and legal process. In: Buzawa ES, Buzawa CG, eds. Domestic Violence: The Changing Criminal Justice Response. Westport, Conn: Auburn House; 1992:227-252.
Klein AR. Re-abuse in a population of court-restrained male batterers after two years: development of a predictive model. In: Buzawa ES, Buzawa CG, eds. Do Arrests and Restraining Orders Work? Thousand Oaks, Calif: SAGE Publications; 1996:192-213.
Harrell A, Smith B, Newmark L. Court Processing and the Effects of Restraining Orders for Domestic Violence Victims. Washington, DC: The Urban Institute; 1993.
Horton AL, Simonidis KM, Simonidis LL. Legal remedies for spousal abuse: victim characteristics, expectations, and satisfaction.  J Fam Viol.1987;2:265-279.
Carlson MJ, Harris SD, Holden GW. Protective orders and domestic violence: risk factors for re-abuse.  J Fam Viol.1999;14:205-226.
Bachman R, Saltzman LE. Violence Against Women: Estimates From the Redesigned Survey. Washington, DC: US Dept of Justice; 1995.
Wolf ME, Kernic MA, Holt, VL, Rivara FP. Do protection orders work?  Am J Epidemiol.2000;151(suppl):S53.
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